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Stochastic processes constrain the within and between host evolution of influenza virus

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Cite this article as: eLife 2018;7:e35962 doi: 10.7554/eLife.35962

Abstract

The evolutionary dynamics of influenza virus ultimately derive from processes that take place within and between infected individuals. Here we define influenza virus dynamics in human hosts through sequencing of 249 specimens from 200 individuals collected over 6290 person-seasons of observation. Because these viruses were collected from individuals in a prospective community-based cohort, they are broadly representative of natural infections with seasonal viruses. Consistent with a neutral model of evolution, sequence data from 49 serially sampled individuals illustrated the dynamic turnover of synonymous and nonsynonymous single nucleotide variants and provided little evidence for positive selection of antigenic variants. We also identified 43 genetically-validated transmission pairs in this cohort. Maximum likelihood optimization of multiple transmission models estimated an effective transmission bottleneck of 1–2 genomes. Our data suggest that positive selection is inefficient at the level of the individual host and that stochastic processes dominate the host-level evolution of influenza viruses.

https://doi.org/10.7554/eLife.35962.001

Introduction

The rapid evolution of influenza viruses has led to reduced vaccine efficacy and the continuing emergence of novel strains. Broadly speaking, evolution is the product of deterministic processes, such as selection, and stochastic processes, such as genetic drift and migration (Kouyos et al., 2006). The global evolution of influenza A virus (IAV) is dominated by the positive selection of novel antigenic variants that subsequently seed annual epidemics in the Northern and Southern hemisphere (Rambaut et al., 2008). These global patterns are contrasted by results from whole genome sequencing of viruses on more local scales, which suggest the importance of stochastic processes such as strain migration and within-clade reassortment (Nelson et al., 2006). While it is clear that influenza’s evolutionary dynamics differ across scales (Nelson and Holmes, 2007; Holmes, 2009), the relative roles of deterministic and stochastic forces in viral evolution within and between acutely infected individuals remain unclear.

It is now feasible to efficiently sequence patient-derived isolates at sufficient depth of coverage to define the genetic diversity and dynamics of virus evolution within individual hosts (Kao et al., 2014). Studies of IAV populations in animal and human systems suggest that most intrahost single nucleotide variants (iSNV) are rare and that intrahost populations are subject to strong purifying selection (Rogers et al., 2015; Murcia et al., 2010; Iqbal et al., 2009; Poon et al., 2016; Dinis et al., 2016; Debbink et al., 2017). Positive selection is common in cell culture (Doud et al., 2017; Archetti and Horsfall, 1950; Foll et al., 2014), and has been observed in experimental infections in swine (Illingworth et al., 2014). However, it has only been well documented within human hosts in the extreme cases of drug resistance (Gubareva et al., 2001; Ghedin et al., 2011; Rogers et al., 2015), and long-term infection of immunocompromised individuals (Xue et al., 2017). Indeed, we and others have been unable to identify evidence for positive selection in naturally-occurring, acute human infections (Debbink et al., 2017; Dinis et al., 2016), and its relevance to within-host processes is unclear.

Despite limited evidence for positive selection, novel mutations do arise within hosts and some will clearly be positively selected. Their potential for subsequent spread through host populations is heavily dependent on the size of the transmission bottleneck (Alizon et al., 2011; Zwart and Elena, 2015). If the transmission bottleneck is sufficiently wide, low frequency variants can plausibly be transmitted and spread efficiently through host populations (Geoghegan et al., 2016). While experimental infections of ferrets suggest a very narrow transmission bottleneck (Varble et al., 2014; Wilker et al., 2013), studies of equine influenza support a bottleneck wide enough to allow transmission of rare iSNV (Hughes et al., 2012; Murcia et al., 2010). The only available genetic study of influenza virus transmission in humans estimated a remarkably large transmission bottleneck, allowing for transmission of 100–200 genomes (Poon et al., 2016; Sobel Leonard et al., 2017).

Here, we use next generation sequencing of within-host influenza virus populations to define the evolutionary dynamics of IAV within and between human hosts. We apply a benchmarked analysis pipeline to identify iSNV and to characterize the genetic diversity of H3N2 and H1N1 populations collected over five post-pandemic seasons from individuals enrolled in a prospective household study of influenza. We find that intrahost populations are dynamic and constrained by genetic drift and purifying selection. In our study, positive selection rarely amplifies a beneficial de novo variant to a frequency greater than 2%. Contrary to what has been previously reported for human influenza transmission (Poon et al., 2016), but consistent with what has been observed in many other viruses with distinct modes of transmission (Zwart and Elena, 2015; McCrone and Lauring, 2018), we identify a very tight effective transmission bottleneck that limits the transmission of low-frequency variants.

Results

We used next generation sequencing to characterize influenza virus populations collected from individuals enrolled in the Household Influenza Vaccine Effectiveness (HIVE) study (Monto et al., 2014; Ohmit et al., 2013; Ohmit et al., 2015; Ohmit et al., 2016; Petrie et al., 2013), a community-based cohort that enrolls 213–340 households of 3 or more individuals in Southeastern Michigan each year (Table 1). These households are followed prospectively from October to April, with symptom-triggered collection of nasal and throat swab specimens for identification of respiratory viruses by RT-PCR (see Materials and methods). In contrast to case-ascertained studies, which identify households based on an index case who seeks medical care, the HIVE study identifies symptomatic individuals regardless of illness severity. In the first four seasons of the study (2010–2011 through 2013–2014), respiratory specimens were collected 0–7 days after illness onset. Beginning in the 2014–2015 season, each individual provided two samples, a self-collected specimen at the time of symptom onset and a clinic-collected specimen obtained 0–7 days later. Each year, 59–69% of individuals had self-reported or confirmed receipt of that season’s vaccine prior to local circulation of influenza virus.

Table 1
Influenza viruses over five seasons in a household cohort
https://doi.org/10.7554/eLife.35962.002
2010–20112011–20122012–20132013–20142014–2015
Households328213321232340
Participants1441943142610491431
Vaccinated, n (%)*934 (65)554 (59)942 (66)722 (69)992 (69)
IAV Positive Individuals86236948166
H1N12613470
H3N25822661166
IAV Positive Households
Two individuals1329723
Three individuals523311
Four individuals--124
High Quality NGS Pairs§412639
  1. *Self reported or confirmed receipt of vaccine prior to the specified season.

    †RT-PCR confirmed infection.

  2. ‡Households in which two individuals were positive within 7 days of each other. In cases of trios and quartets, the putative chains could have no pair with onset > 7 days apart.

    §Samples with > 103 genome copies per µl of transport medium, adequate amplification of all eight genomic segments, and average sequencing coverage > 103 per nucleotide.

Over five seasons and nearly 6290 person-seasons of observation, we identified 77 cases of influenza A/H1N1pdm09 infection and 313 cases of influenza A/H3N2 infection (Table 1). Approximately half of the cases (n = 166) were identified in the 2014–2015 season, in which there was an antigenic mismatch between the vaccine and circulating strains (Flannery et al., 2016). All other seasons were antigenically matched. Individuals within a household were considered an epidemiologically linked transmission pair if they were both positive for the same subtype of influenza virus within 7 days of each other. Several households had 3 or four symptomatic cases within this one-week window, suggestive of longer chains of transmission (Table 1).

Within-host populations have low genetic diversity

We processed all specimens for viral load quantification and next generation sequencing. Viral load measurements (genome copies per µl) were used for quality control in variant calling, which we have shown is highly sensitive to input titer (McCrone and Lauring, 2016) (Figure 1A). Accordingly, we report data on 249 high quality specimens from 200 individuals, which had a viral load of >103 copies per microliter of transport media, adequate RT-PCR amplification of all eight genomic segments, and an average read coverage of >103 across the genome (Table 1, Figure 1—figure supplement 1).

Figure 1 with 4 supplements see all
Within-host diversity of IAV populations.

(A) Boxplots (median, 25th and 75th percentiles, whiskers extend to most extreme point within median ±1.5 x IQR) of the number of viral genomes per microliter transport media stratified by day post symptom onset. Notches represent the approximate 95% confidence interval of the median. (B) Boxplots (median, 25th and 75th percentiles, whiskers extend to most extreme point within median ±1.5 x IQR) of the number of iSNV in 249 high quality samples stratified by day post symptom onset. (C) The number of iSNV in each isolate stratified by vaccination status. The red lines indicate the median. (D) Location of all identified iSNV in the influenza A genome. Mutations are colored nonsynonymous (blue) and synonymous (gold) relative to that sample’s consensus sequence. Mutations are considered nonsynonymous if they are nonsynonymous in any known influenza open reading frame. Triangles signify mutations that were found in more than one individual in a given season.

https://doi.org/10.7554/eLife.35962.003

We identified intrahost single nucleotide variants (iSNV) using our empirically validated analysis pipeline (McCrone and Lauring, 2016). Our approach relies heavily on the variant caller DeepSNV, which uses a clonal plasmid control to distinguish between true iSNV and errors introduced during sample preparation and/or sequencing (Gerstung et al., 2012). Given the diversity of influenza viruses that circulate locally each season, there were a number of instances in which our patient-derived samples had mutations that were essentially fixed (>0.95 frequency) relative to the clonal control. DeepSNV is unable to estimate an error rate for the control or reference base at these positions. We therefore performed an additional benchmarking experiment to identify a threshold for majority iSNV at which we could correctly infer whether or not the corresponding minor allele was also present (see Materials and methods). We found that we could correctly identify a minor allele at a frequency of ≥2% at such sites. We therefore report data on iSNV present at frequencies between 2% and 98%. As expected, this threshold improved the specificity of our iSNV identification and decreased our sensitivity to detect variants below 5% compared to our initial validation experiment (McCrone and Lauring, 2016), which did not employ a frequency threshold (Supplementary file 1).

Consistent with our previous studies of natural infections and those of others, we found that the within-host diversity of seasonal influenza A virus (IAV) populations is low (Dinis et al., 2016; Debbink et al., 2017; Sobel Leonard et al., 2016; McCrone and Lauring, 2016). Two hundred forty-three out of the 249 samples had fewer than 10 minority iSNV (median 2, IQR 1–3). There were six samples with greater than 10 minority iSNV. In 3 of these cases, the frequency distribution of minority iSNVs was bimodal, suggesting that the iSNV were linked and that the samples represented mixed infections. Consistent with this hypothesis, putative genomic haplotypes based on these minority iSNV clustered with distinct isolates on phylogenetic trees (Figure 1—figure supplements 2 and 3), and these samples were removed from subsequent analysis. While viral shedding was well correlated with days post symptom onset (Figure 1A), the number of minority iSNV identified was not affected by the day of infection, viral load, subtype, or vaccination status (Figure 1B and C, and Figure 1—figure supplement 4). Single nucleotide variants were distributed evenly across the genome (Figure 1D).

Minority variants were rarely shared among multiple individuals. Ninety-eight percent of minority iSNV were only found once, 2.3% were found in two individuals, and no minority iSNV were found in three or more individuals. The low level of shared diversity suggests that within-host populations explore distinct regions of sequence space with little evidence for parallel evolution. Of the 12 minority iSNV that were found in multiple individuals (triangles in Figure 1D), three were nonsynonymous and none were present in antigenic epitopes. The vast majority of minority variants were rare (frequency 0.02–0.07) (Figure 2A). The ratio of nonsynonymous to synonymous variants was 0.75, and given the excess of nonsynonymous sites across the genome and within the HA gene, these data suggest significant purifying selection within hosts.

Figure 2 with 1 supplement see all
Within-host dynamics of IAV.

(A) Histogram of within-host iSNV frequency in 249 high quality samples. Bin width is 0.05 beginning at 0.02. As in Figure 1, mutations were classified as nonsynonymous (blue) if they were nonsynonymous in any known influenza reading frame. Synonymous mutations are gold. (B) The within-host frequency of nonsynonymous mutations in HA stratified by whether or not they are in known antigenic sites (p=0.46 Wilcoxon rank sum). (C) The global frequency of putative antigenic minority iSNV identified in our cohort that have circulated at frequencies above 5% globally since their time of collection. Each variant is labeled according the H3 numbering scheme. The dashed line indicates when samples were collected. Frequency traces are faded prior to the collection date. (D) Timing of sample collection for 43 paired longitudinal samples relative to day of symptom onset. Of the 49 total, 43 pairs had minority iSNV present in either sample. (E) The change in frequency over time for minority iSNV identified for the paired samples in (A). Nonsynonymous and synonymous iSNV are plotted separately. Mutations are colored according to whether they were detected in both isolates (blue), detected only the first isolate (red), or detected only in the second isolate (yellow). The threshold of detection was 2%. The arrows indicate mutations in known antigenic sites.

https://doi.org/10.7554/eLife.35962.012

iSNV within and outside antigenic epitopes have similar frequencies

Although the full range of the H3 antigenic sites have not been functionally defined, it is estimated that 131 of the 329 amino acids in HA1 lie in or near these sites (Lee and Chen, 2004). Although we identified 16 minority nonsynonymous iSNV in these regions (Supplementary file 2), mutations in antigenic epitopes were not present at significantly higher within-host frequencies than other nonsynonymous mutations in HA (Figure 2B). Five putative antigenic variants were in positions that have been experimentally shown to differ among antigenically drifted viruses (Smith et al., 2004; Wiley et al., 1981), and two (193S and 189N) lie in the ‘antigenic ridge’ that is a major contributor to antigenic drift (Koel et al., 2013). Five of the iSNV in antigenic epitopes have been detected as minority variants above 5% at the global level since the time of isolation (62G, 128A, 193S, 262N, and 307R) with one (62G) seemingly increasing in global frequency (Figure 2C). An additional within host variant (307R) has dominated the population for the past 12 years (Neher and Bedford, 2015). We identified one putative H1N1 antigenic variant (208K in antigenic site Ca) (Caton et al., 1982; Xu et al., 2010). In total, putative antigenic variants account for 1.0–2.8% of minority iSNV identified and were found in 3.0–7.1% of infections. None of these iSNV were shared among multiple individuals.

Within host populations are dynamic

We next examined the changes in within-host virus diversity over time in 49 individuals who provided paired longitudinal samples during the 2014/2015 season (Figure 2D). We found that there was very little change in iSNV frequency in populations sampled twice on the same day (R2 = 0.982, Figure 2E and Figure 2—figure supplement 1A), and the concordance of same day samples suggests that our sampling procedure is reproducible. An analysis of sequence data from replicate libraries indicates that our estimates of iSNV frequency are precise (Figure 2—figure supplement 1B). In samples separated by at least a day, only 57% iSNV found in the first sample persisted in the second, above the 2% limit of detection. Additionally, the majority of iSNV (68%) found in the second sample were either new or previously present below the 2% limit of detection. Taken together, these data suggest that the population present in the upper respiratory tract is highly dynamic while maintaining a stable consensus. Of note, 6 iSNV in our longitudinal data set lay within antigenic epitopes (arrows in Figure 2E). Their dynamics are similar to those of other nonsynonymous iSNV and synonymous iSNV, suggesting that most mutations change in frequency due to stochastic as opposed to selective processes. Together with our prior work (Debbink et al., 2017), these data suggest that the positive selection of novel variants within hosts is inefficient and rarely amplifies a newly generated variant to a frequency greater than 2%.

Identification of forty-three transmission pairs

Our within-host data suggest that newly arising iSNV with positive fitness effects are likely to be present at low frequencies (<2%) during an acute infection. The maintenance of these mutations in host populations is therefore highly dependent on the transmission bottleneck. We analyzed virus populations from 85 households with concurrent infections to quantify the level of shared viral diversity and to estimate the size of the IAV transmission bottleneck (Table 1). Because epidemiological linkage does not guarantee that concurrent cases constitute a transmission pair (Petrie et al., 2017), we used stringent criteria to eliminate individuals in a household with co-incident community acquisition of distinct viruses. We considered all individuals in a household with symptom onset within a 7 day window to be epidemiologically linked. The donor in each putative pair was defined as the individual with the earlier onset of symptoms. We discarded a transmission event if there were multiple possible donors with the same day of symptom onset. Donor and recipients were not allowed to have symptom onset on the same day, unless the individuals were both index cases for the household. In these six instances, we analyzed the data for both possible donor-recipient directionalities. Based on these criteria, our cohort had 124 putative household transmission events over five seasons (Table 1). Of these, 52 pairs had samples of sufficient quality for reliable identification of iSNV from both individuals.

We next used sequence data to determine which of these 52 epidemiologically linked pairs represented true household transmission events as opposed to coincident community-acquired infections. We measured the genetic distance between influenza populations from each household pair by L1-norm and compared these distances to those of randomly assigned community pairs within each season (Figure 3A, see also trees in Figure 1—figure supplements 2 and 3). While the L1-norm of a pair captures differences between the populations at all levels, in our cohort, it was largely driven by differences at the consensus level. We only considered individuals to be a true transmission pair if they had a genetic distance below the 5th percentile of the community distribution of randomly assigned pairs (Figure 3A). Forty-seven household transmission events met this criterion (Figure 3B). Among these 47 sequence-validated transmission pairs, three had no iSNV in the donor and one additional donor appeared to have a mixed infection. These four transmission events were removed from our bottleneck analysis, as donors without iSNV are uninformative and mixed infections violate model assumptions of site independence (see Materials and methods). We estimated the transmission bottleneck in the remaining 43 high-quality pairs (37 H3N2, 6 H1N1, Figure 3B).

Figure 3 with 1 supplement see all
Between host dynamics of IAV.

(A) The distribution of pairwise L1-norm distances for household (blue) and randomly-assigned community (gold) pairs. The bar heights are normalized to the height of the highest bar for each given subset (47 for household, 1592 for community). The red line represents the 5th percentile of the community distribution. (B) Timing of symptom onset for 52 epidemiologically linked transmission pairs. Days of symptom onset for both donor and recipient individuals are indicated by black dots. Dashed lines represent pairs that were removed due to abnormally high genetic distance between isolates, see (A). (C) The frequency of donor iSNV in both donor and recipient samples. Frequencies below 2% and above 98% were set to 0% and 100% respectively. (D) The presence-absence model fit compared with the observed data. The x-axis represents the frequency of donor iSNV with transmitted iSNV plotted along the top and nontransmitted iSNV plotted along the bottom. The line represents the predicted probability of transmission given by the presence-absence model with a mean bottleneck of 1.68. The shaded regions represent the 95% confidence interval. Black points on the plot represent the probability of transmission estimated as the proportion of iSNV transmitted within a sliding window of width 5% and a step of 1%. The error bars represent the 95% confidence interval and were derived from a binomial distribution as in (Sobel Leonard et al., 2017). Only those windows with more than 5 iSNV are plotted. Blue curve shows the probability of transmission at a given frequency given a bottleneck size of 10 in the presence-absence model. (E) The beta-binomial model fit. Similar to (D), except the predicted outcomes are the based on a beta-binomial model using a mean bottleneck of 1.75. Blue curve shows the probability of transmission at a given frequency given a bottleneck size of 10 in the beta-binomial model.

https://doi.org/10.7554/eLife.35962.019

A transmission bottleneck restricts the amount of genetic diversity that is shared by both members of a pair. We found that few minority iSNV where polymorphic in both the donor and recipient populations (Figure 3C). Minority iSNV in the donor were either absent or fixed in the recipient (top and bottom of plot). The lack of shared polymorphic sites (which would lie in the middle of the plot in Figure 3C) suggests a stringent effective bottleneck in which only one allele is passed from donor to recipient.

Estimation of the transmission bottleneck

We applied a simple presence-absence model to quantify the effective transmission bottleneck in our cohort. The presence-absence model considers only whether or not a donor allele is present or absent in the recipient sample. Under this model, transmission is a neutral, random sampling process, and the probability of transmission is simply the probability that the iSNV will be included at least once in the sample given its frequency in the donor and the sample size, or bottleneck. We estimated a distinct bottleneck for each transmission pair and assumed these bottlenecks followed a zero-truncated Poisson distribution. Maximum likelihood optimization determined that a mean bottleneck of 1.68 (lambda = 1.15; 0.49–2.14, 95% CI) best described the data. This distribution indicates that the majority of bottlenecks are 1 and that 95% of bottlenecks are expected to be ≤3. One outlier was a single transmission pair with an estimated bottleneck of >200; this pair had two minority iSNV present at very low frequencies in both donor and recipient. There were no apparent differences between H3N2 and H1N1 pairs. The model fit was evaluated by comparing the probability of transmission predicted by the model with that estimated from the data using a sliding window (Figure 3D). The predicted transmission probabilities capture the underlying trend in the data, which suggests that transmission dynamics can be appropriately modeled using a selectively neutral, stringent bottleneck.

Because the presence-absence model can underestimate the transmission bottleneck in some circumstances (Sobel Leonard et al., 2017), we also applied a beta binomial model, which Leonard et al. have used to account for the stochastic dynamics of transmitted variants. This model allows for a limited amount of time-independent genetic drift within the recipient (Sobel Leonard et al., 2017), and we modified it to also account for our benchmarked sensitivity for rare variants (Supplementary file 1, Current Pipeline). For all donor-derived iSNV that were absent in the recipient, we estimated the likelihood that these variants were transmitted but either drifted below our level of detection or drifted below 10% and were missed by our variant identification. Despite the relaxed assumptions provided by this modified beta binomial model, maximum likelihood estimation only marginally increased the average bottleneck size (mean 1.75: lambda 1.25; 0.55–2.28, 95% CI) relative to the simpler presence-absence model. We also fit a long-tailed, discrete distribution based on the log-normal. As expected, this analysis resulted in a slightly wider distribution with a mode of 1, a 95th percentile of 11 and a 97.5th percentile of 21. Although the beta-binomial again reproduced the trend in the data (Figure 3E), the fit was not better than that of the presence-absence model (AIC 80.2 for beta-binomial compared to 74.5 for the presence-absence model), suggesting the stochastic loss of iSNV in the recipient is not a major contributing factor to the transmission dynamics observed here. As in the presence-absence model, there was no apparent stratification in bottleneck size based on vaccination status, donor or recipient age, or influenza A subtype (Supplementary file 3).

Because our bottleneck estimates were much lower than what has previously been reported for human influenza (Poon et al., 2016), we investigated the impact that our simplifying assumptions could have on our results. In particular, the presence-absence model assumes perfect detection of variants in donor and recipient, and it can therefore underestimate the size of a bottleneck in the setting of donor-derived variants that are transmitted but not detected in the recipient. These ‘false negative’ variants can occur when the frequency of an iSNV drifts below the level of detection (e.g. 2% frequency) or when the sensitivity of sequencing is less than perfect for variants at that threshold (e.g. 15% sensitivity for variants at a frequency 2–5%). To determine the impact of sequencing sensitivity and specificity on our bottleneck estimates, we re-called variants using our original pipeline without the 2% frequency cut-off. As shown in Supplementary file 1, this increases the sensitivity of iSNV detection in the 1–5% frequency range, and also the number of false positive variant calls (McCrone and Lauring, 2016). This analysis only slightly increased average transmission bottleneck to 2.11 (lambda = 1.75; 0.87–2.95, 95% CI), and indicates that our results are not biased by the added stringency used in the initial analysis. Interestingly, the inclusion of these low-frequency iSNV reduced model fit and led to an overestimation of the probability of transmitting rare iSNV (Figure 3—figure supplement 1). These low-frequency bins are likely dominated by iSNV that were not present at the time of transmission either because they are sequencing artifacts or underwent stochastic extinction in the donor prior to transmission.

Discussion

We characterized influenza virus dynamics in human hosts through sequencing of 249 specimens from 200 individuals collected over 6290 person-seasons of observation. In our study, we find that acute influenza infections are characterized by low diversity, limited positive selection, and tight transmission bottlenecks. Because we used viruses collected over five influenza seasons from individuals enrolled in a prospective household cohort, these dynamics are likely to be broadly representative of many seasonal influenza infections in temperate regions. Our results are further strengthened by the use of a validated sequence analysis pipeline and models that are robust to the underlying assumptions. The observed within-host dynamics and the tight effective transmission bottleneck suggest that stochastic processes, such as genetic drift, dominate influenza virus evolution at the level of individual hosts. This stands in contrast to the prominent role of positive selection in the global evolution of seasonal influenza.

Contrary to previous studies, which have found signatures of adaptive evolution in infected hosts (Gubareva et al., 2001; Rogers et al., 2015; Ghedin et al., 2011; Sobel Leonard et al., 2016), we have found only limited evidence of positive selection during acute infection. Previous reports have relied on infections in which selective pressures are likely to be particularly strong (e.g. due to drug treatment or infection with a poorly adapted virus), or in which the virus has replicated within a host for an extended period of time (Xue et al., 2017). Under these conditions, it is plausible that positively selected alleles reach levels of detection. We suggest that these are important and informative exceptions to dynamics defined here, in which positive selection is rarely strong enough to drive a new mutation to a frequency above 2% over the course of several days. Our findings are consistent with the observed global rates of influenza evolution and with epidemiological studies that have shown limited antigenic selection over the course of local epidemics (Nelson et al., 2006; Bedford et al., 2015).

We used both a simple presence-absence model and a more complex beta binomial model to estimate an extremely tight transmission bottleneck. The estimation of a small bottleneck is due to the fact that at most polymorphic sites in the donor, only one allele (iSNV) was found in the recipient. While our methods for variant calling may be more conservative than those used in similar studies, we found that relaxing our variant calling criteria only led to the inclusion of false positive variants and did not significantly inflate our estimates. Furthermore, the beta binomial model accounts for false negative iSNV (i.e. variants that are transmitted but not detected in the donor), which can lead to underestimated transmission bottlenecks (Sobel Leonard et al., 2017). Our formulation of this model incorporates empirically determined sensitivity and specificity metrics to account for both false negative iSNV and false positive iSNV (McCrone and Lauring, 2016). Finally, if rare, undetected, iSNV were shared between linked individuals, we would expect to see transmission of more common iSNV (frequency 5–10%), which we can detect with high sensitivity. In our data, the transmission probability iSNVs > 5% frequency in the donor were also well predicted by small bottleneck size (Figure 3D and E).

Although the size of our transmission bottleneck is consistent with estimates obtained for other viruses and in experimental animal models of influenza (Zwart and Elena, 2015; Varble et al., 2014), it differs substantially from the only other study of bottlenecks in natural human infection (Poon et al., 2016; Sobel Leonard et al., 2017). While there are significant differences in the design and demographics of the cohorts, the influenza seasons under study, and sequencing methodology (Kugelman et al., 2017), the bottleneck size estimates are fundamentally driven by the amount of viral diversity shared among individuals in a household. Importantly, and as in Poon et al., we used both epidemiologic linkage and the genetic relatedness of viruses in households to define transmission pairs and to exclude confounding from the observed background diversity in the community. We find that household transmission pairs and randomly assigned community pairs had distinct patterns of shared consensus and minority variant diversity, a pattern that is distinct from the observations of Poon et al. An unexplained aspect of their study is that rare iSNV were frequently shared by randomly selected individuals, and more common ones were not (Poon et al., 2016).

Accurately modeling and predicting influenza virus evolution requires a thorough understanding of the virus’ population structure. Some models have assumed a large intrahost population and a relatively loose transmission bottleneck (Geoghegan et al., 2016; Russell et al., 2012; Peck et al., 2015). Here, adaptive iSNV can rapidly rise in frequency and low frequency variants can have a high probability of transmission. In such a model, it would be possible for an emerging virus to become more transmissible or a seasonal virus to evolve resistance to vaccine-induced immunity over a short transmission chain (Herfst et al., 2012; Russell et al., 2012). Although the dynamics of emergent avian influenza and human adapted seasonal viruses likely differ (Petrova and Russell, 2017), our work suggests that fixation of multiple mutations over the course of a single acute infection is unlikely.

While it may seem counterintuitive that influenza evolution is dominated by stochasticity on local scales and positive selection on global scales, these models are certainly not in conflict. We have deeply sequenced 332 intrahost populations from 283 individuals collected over more than 11,000 person-seasons of observation and only identified a handful of minority antigenic variants with limited evidence for positive selection (this work and [Debbink et al., 2017]). Importantly, our data suggest that even if selection acts below our level of detection, such rare variants are unlikely to transmit. Given the size of the estimated bottleneck, the probability of transmission is approximately 1.7% for a variant at 1% frequency and 3.3% for a variant at 2% frequency. However, with several million infected individuals each year, even inefficient processes and rare events at the scale of individual hosts are likely to occur at a reasonable frequency on a global scale.

Materials and methods

Description of the cohort

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The HIVE cohort (Monto et al., 2014; Ohmit et al., 2013, 2015, 2016; Petrie et al., 2013), established at the UM School of Public Health in 2010, enrolled and followed households of at least three individuals with at least two children < 18 years of age; households were then followed prospectively throughout the year for ascertainment of acute respiratory illnesses. Study participants were queried weekly about the onset of illnesses meeting our standard case definition (two or more of: cough, fever/feverishness, nasal congestion, sore throat, body aches, chills, headache if ≥3 years old; cough, fever/feverishness, nasal congestion/runny nose, trouble breathing, fussiness/irritability, decreased appetite, fatigue in <3 years old), and the symptomatic participants then attended a study visit at the research clinic on site at UM School of Public Health for sample collection. For the 2010–2011 through 2013–2014 seasons, a combined nasal and throat swab (or nasal swab only in children < 3 years of age) was collected at the onsite research clinic by the study team. Beginning with the 2014–2015 seasons, respiratory samples were collected at two time points in each participant meeting the case definition; the first collection was a self- or parent-collected nasal swab collected at illness onset. Subsequently, a combined nasal and throat swab (or nasal swab only in children < 3 years of age) was collected at the onsite research clinic by the study team. Families with very young children (<3 years of age) were followed using home visits by a trained medical assistant.

Active illness surveillance and sample collection for cases were conducted October through May and fully captured the influenza season in Southeast Michigan in each of the study years. Data on participant, family and household characteristics, and on high-risk conditions were additionally collected by annual interview and review of each participant’s electronic medical record. In the current cohort, serum specimens were also collected twice yearly during fall (November-December) and spring (May-June) for serologic testing for antibodies against influenza.

This study was approved by the Institutional Review Board of the University of Michigan Medical School. Adults provided written informed consent for participation for themselves and their children; children 7–17 years provided oral assent.

Identification of influenza virus

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Respiratory specimens were processed daily to determine laboratory-confirmed influenza infection. Viral RNA was extracted (Qiagen QIAamp Viral RNA Mini Kit, Germantown, MD) and tested by RT-PCR for universal detection of influenza A and B. Samples with positive results by the universal assay were then subtyped to determine A(H3N2), A(H1N1), A(pH1N1) subtypes and B(Yamagata) and B(Victoria) lineages. We used primers, probes and amplification parameters developed by the Centers for Disease Control and Prevention Influenza Division for use on the ABI 7500 Fast Real-Time PCR System platform. An RNAseP detection step was run for each specimen to confirm specimen quality and successful RNA extraction.

Quantification of viral load

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Quantitative reverse transcription polymerase chain reaction (RT-qPCR) was performed on 5 μl RNA from each sample using CDC RT-PCR primers InfA Forward, InfA Reverse, and InfA probe, which bind to a portion of the influenza M gene (CDC protocol, 28 April 2009). Each reaction contained 5.4 μl nuclease-free water, 0.5 μl each primer/probe, 0.5 μl SuperScript III RT/Platinum Taq mix (Invitrogen111732, Carlsbad, CA) 12.5 μl PCR Master Mix, 0.1 μl ROX, 5 μl RNA. The PCR master mix was thawed and stored at 4°C, 24 hr before reaction set-up. A standard curve relating copy number to Ct value was generated based on 10-fold dilutions of a control plasmid run in duplicate.

Illumina library preparation and sequencing

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We amplified cDNA corresponding to all eight genomic segments from 5 μl of viral RNA using the SuperScript III One-Step RT-PCR Platinum Taq HiFi Kit (Invitrogen 12574). Reactions consisted of 0.5 μl Superscript III Platinum Taq Mix, 12.5 μl 2x reaction buffer, 6 μl DEPC water, and 0.2 μl of 10 μM Uni12/Inf1, 0.3 μl of 10 μM Uni12/Inf3, and 0.5 μl of 10 μM Uni13/Inf1 universal influenza A primers (Zhou et al., 2009). The thermocycler protocol was: 42 ˚C for 60 min then 94 ˚C for 2 min then 5 cycles of 94 ˚C for 30 s, 44 ˚C for 30 s, 68 ˚C for 3 min, then 28 cycles of 94 ˚C for 30 s, 57 ˚C for 30 s, 68 ˚C for 3 min. Amplification of all eight segments was confirmed by gel electrophoresis, and 750 ng of each cDNA mixture were sheared to an average size of 300 to 400 bp using a Covaris (Woburn, MA) S220 focused ultrasonicator. Sequencing libraries were prepared using the NEBNext Ultra DNA library prep kit (NEB E7370L), Agencourt AMPure XP beads (Beckman Coulter A63881, Indianapolis, IN), and NEBNext multiplex oligonucleotides for Illumina (NEB E7600S, Ipswich, MA). The final concentration of each barcoded library was determined by Quanti PicoGreen dsDNA quantification (ThermoFisher Scientific, Waltham, MA), and equal nanomolar concentrations were pooled. Residual primer dimers were removed by gel isolation of a 300–500 bp band, which was purified using a GeneJet Gel Extraction Kit (ThermoFisher Scientific). Purified library pools were sequenced on an Illumina HiSeq 2500 with 2 × 125 nucleotide paired end reads. All raw sequence data have been deposited at the NCBI sequence read archive (BioProject Accession number: PRJNA412631). PCR amplicons derived from an equimolar mixture of eight clonal plasmids, each containing a genomic segment of the circulating strain were processed in similar fashion and sequenced on the same HiSeq flow cell as the appropriate patient derived samples. These clonally derived samples served as internal controls to improve the accuracy of variant identification and control for batch effects that confound sequencing experiments.

Identification of iSNV

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Intrahost single nucleotide variants were identified in samples that had greater than 103 genomes/μl and an average coverage >1000 x across the genome. Variants were identified using DeepSNV and scripts available at https://github.com/lauringlab/variant_pipeline as described previously (McCrone and Lauring, 2016; copy archived at https://github.com/elifesciences-publications/variant_pipeline) with a few minor and necessary modifications. Briefly, reads were aligned to the reference sequence (H3N2 2010–2011 and 2011–2012: GenBank CY121496-503, H3N2 2012–2013:GenBank KJ942680-8, H3N2 2014–2015: Genbank CY207731-8, H1N1 GenBank: CY121680-8) using Bowtie2 (35). Duplicate reads were then marked and removed using Picard (http://broadinstitute.github.io/picard/). We identified putative iSNV using DeepSNV. Bases with phred <30 were masked. Minority iSNV (frequency <50%) were then filtered for quality using our empirically determined quality thresholds (p-value<0.01 DeepSNV, average mapping quality >30, average Phred > 35, average read position between 31 and 94). To control for RT-PCR errors in samples with lower input titers, all isolates with titers between 103 and 105 genomes/μl were processed and sequenced in duplicate. Only iSNV that were found in both replicates were included in down stream analysis. The frequency of the variant in the replicate with higher coverage at the iSNV location was assigned as the frequency of the iSNV. Finally, any SNV with a frequency below 2% was discarded.

Given the diversity of the circulating strain in a given season, there were a number of cases in which isolates contained mutations that were essentially fixed (>95%) relative to the plasmid control. Often in these cases, the minor allele in the sample matched the major allele in the plasmid control. We were, therefore, unable to use DeepSNV in estimating the base specific error rate at this site for these minor alleles and required an alternative means of discriminating true and false minority iSNV. To this end we applied stringent quality thresholds to these putative iSNV and implemented a 2% frequency threshold. In order to ensure we were not introducing a large number of false positive iSNV into our analysis, we performed the following experiment. Perth (H3N2) samples were sequenced on the same flow cell as both the Perth and Victoria (H3N2) plasmid controls. Minority iSNV were identified using both plasmid controls. This allowed us to identify rare iSNV at positions in which the plasmid controls differed both with and without the error rates provided by DeepSNV. We found that at a frequency threshold of 2% the methods were nearly identical (NPV of 1, and PPV of 0.94 compared to DeepSNV).

Overview of models used for estimating the transmission bottleneck

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We model transmission as a simple binomial sampling process (Sobel Leonard et al., 2017). In our first model, we assume any transmitted iSNV, no matter the frequency, will be detected in the recipient. In the second, we relax this assumption and account for false negative iSNV in the recipient. To include the variance in the transmission bottlenecks between pairs we use maximum likelihood optimization to fit the average bottleneck size assuming the distribution follows a zero-truncated Poisson distribution.

Presence/Absence model

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The presence/absence model makes several assumptions. We assume perfect detection of all transmitted iSNV in the recipient. For each donor iSNV, we measure only whether or not the variant is present in the recipient. Any iSNV that is not found in the recipient is assumed to have not been transmitted. We also assume the probability of transmission is determined only by the frequency of the iSNV in the donor at the time of sampling (regardless of how much time passes between sampling and transmission). The probability of transmission is simply the probability that the iSNV is included at least once in a sample size equal to the bottleneck. Finally, we assume all genomic sites are independent of one another. For this reason, we discarded the one case where the donor was likely infected by two strains, as the iSNV were certainly linked.

Because the presence/absence model is unaware of the frequency of alleles in the recipient we must track both alleles at each donor polymorphic site.

Let A1 and A2 be alleles in donor j at genomic site i. Let P(A1) be the probability that A1 is the only transmitted allele. There are three possible outcomes for each site. Either only A1 is transmitted, only A2 is transmitted, or both A1 and A2 are transmitted. The probability of only A1 being transmitted given a bottleneck size of Nb is

(1) Pi,j(A1Nb)=p1Nb

where p1 is the frequency of A1 in the donor. In other words, this is simply the probability of only drawing A1 in Nb draws. The probability that only A2 is transmitted is similarly defined.

The probability of both alleles being transmitted is given by

(2) Pi,j(A1,A2Nb)=1-(p1Nb+p2Nb)

where p1 and p2 are the frequencies of the alleles respectively. This is simply the probability of not picking only A1 or only A2 in Nb draws.

This system could easily be extended to cases where there are more than two alleles present at a site; however, that never occurs in our data set.

For ease, we will denote the likelihood of observing the data at a polymorphic site i in each donor j given the bottleneck size Nb as Pi,j(Nb) where Pi,j(Nb) is defined by equation 1 if only one allele is transmitted and equation 2 if two alleles are transmitted.

The log likelihood of a bottleneck of size Nb is given by

(3) LL(Nb)=jiLn(Pi,j)

where i,j refers to the ith polymorphic site in the jth donor. This is the log of the probability of observing the data summed over all polymorphic sites across all donors.

Because the bottleneck size is likely to vary across transmission events, we used maximum likelihood to fit the bottleneck distribution as opposed to fitting a single bottleneck value. Under this model we assumed the bottlenecks were distributed according to a zero-truncated Poisson distribution parameterized by λ. The likelihood of observing the data given a polymorphic site i in donor j and λ is

(4) Pi,j(λ)=Nb=1Pi,j(Nb)P(Nbλ)

where Pi,j(Nb) is defined as above, P(Nbλ) is the probability of drawing a bottleneck of size Nb from a zero-truncated Poisson distribution with a mean of λ1-e-λ. The sum is across all possible Nb defined on [1,). Unless otherwise noted, we only investigated bottleneck sizes up to 100, as initial analyses suggested λ is quite small and the probability of drawing a bottleneck size of 100 from a zero-truncated Poisson distribution with λ=10 is negligible. We follow this convention whenever this sum appears.

The log likelihood of λ for the data set is given by

(5) LL(λ)=jiLn(Nb=1Pi,j(Nb)P(Nbλ))

Beta binomial model

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The Beta binomial model is explained in detail in Leonard et al. (Sobel Leonard et al. 2017). It is similar to the presence/absence model in that transmission is modeled as a simple sampling process; however, it relaxes the following assumptions. In this model, the frequencies of transmitted variants are allowed to change between transmission and sampling according a beta distribution. The distribution is not dependent on the amount of time that passes between transmission and sampling, but rather depends on the size of the founding population (here assumed to equal to Nb) and the number of variant genomes present in founding population k. Note the frequency in the donor is assumed to be the same between sampling and transmission.

The equations below are very similar to those presented by Leonard et al. with one exception. Because we know the sensitivity of our method to detect rare variants based on the expected frequency and the titer, we can include the possibility that iSNV are transmitted but are missed due to poor sensitivity. Because the beta binomial model is aware of the frequency of the iSNV in the recipient, no information is added by tracking both alleles at a genomic site i.

Let pi,jd represent the frequency of the minor allele at position i in the donor of some transmission pair j. Similarly, let pi,jr be the frequency of that same allele in the recipient of the jth transmission pair. Then, as in Leonard et al., the likelihood of some bottleneck Nb for the data at site i in pair j where the minor allele is transmitted is given by

(6) L(Nb)i,j=k=1Nbp_beta(pi,jrk,Nb-k)p_bin(kNb,pi,jd)

Where p_beta is the probability density function for the beta distribution and p_bin is the probability mass function for the binomial distribution.

This is the probability density that the transmitted allele is found in the recipient at a frequency of pi,jr given that the variant was in k genomes in a founding population of size Nb times the probability of drawing k variant genomes in a sample size of Nb and a variant frequency of pi,jd. This is then summed for all possible k where 1kNb.

As in equation 4 the likelihood of a zero truncated Poisson with a mean of λ1-e-λ given this transmitted variants is then given by

(7) L(λ)i,jtransmitted=Nb=1L(Nb)i,jP(Nbλ)

This is simply the likelihood of each Nb weighted by the probability of drawing a bottleneck size of Nb from bottleneck distribution.

In this model, there are three possible mechanisms for a donor iSNV to not be detected in the recipient. (i) The variant was not transmitted. (ii) The variant was transmitted but is present below our limit of detection (2%). (iii) The variant was transmitted and is present above our limit of detection but represents a false negative in iSNV identification.

As in Leonard et al., the likelihood of scenarios (i) and (ii) for a given Nb are expressed as

(8) L(Nb)i,jlost=k=0Nbp_beta_cdf(pi,jr<0.02k,Nbk)p_bin(kNb,pi,jd)

Where p_beta_cdf is the cumulative distribution function for the beta distribution. Note that if k=0 (i.e. the iSNV was not transmitted) then the term reduces to the probability of not drawing the variant in Nb draws.

The likelihood of the variant being transmitted but not detected in the recipient given a bottleneck of Nb is described by

(9) L(Nb)i,jmissed=k=0Nbfe[0.02,0.05,0.1)p_beta_cdf(fe<pi,jr<fe+1k,Nbk)×p_bin(kNb,pi,jd)(FNRTiterr,fe)

This is the likelihood of the variant existing in the ranges [0.02,0.05] or [0.05,0.1] given an initial frequency of k/Nb and a bottleneck size of Nb multiplied by the expected false negative rate (FNR) given the titer of the recipient and the lower frequency bound. We assumed perfect sensitivity for detection of iSNV present above 10%, rounded recipient titers down to the nearest log10 titer (e.g. 103,104, 105) and assumed the entire range [fe,fe+1] has the same sensitivity as the lower bound.

The likelihood of λ for iSNV that are not observed in the recipient is then given by summing equations 8 and 9 across all possible Nb.

(10) L(λ)i,jnontransmitted=Nb=1(L(Nb)i,jlost+L(Nb)i,jmissed)P(Nbλ)

The log likelihood of the total dataset is then determined by summing the log of equations 7 and 10 (as applicable) across all polymorphic sites in each donor. (As before here we sum of Nb within the range [1,100].)

Model comparisons

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In order evaluate the fits of the two transmission models, we calculated the probability of detecting a donor iSNV in the recipient given its frequency in the donor. For each model this reduces to one minus the probability of only detecting the other allele in the recipient summed over all possible bottlenecks. As in fitting the models this infinite sum was approximated by a partial sum up to a bottleneck size of 100. We used a recipient titer of 104 genomes/μl to estimate the effect of sequencing sensitivity for the beta binomial model.

Annotated computer code for all analyses and for generating the figures can be accessed at https://github.com/lauringlab/Host_level_IAV_evolution. (copy archived at https://github.com/elifesciences-publications/Host_level_IAV_evolution)

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Decision letter

  1. Richard A Neher
    Reviewing Editor; University of Basel, Switzerland

In the interests of transparency, eLife includes the editorial decision letter and accompanying author responses. A lightly edited version of the letter sent to the authors after peer review is shown, indicating the most substantive concerns; minor comments are not usually included.

[Editors’ note: a previous version of this study was rejected after peer review, but the authors submitted for reconsideration. The first decision letter after peer review is shown below.]

Thank you for submitting your work entitled "Stochastic processes dominate the within and between host evolution of influenza virus" for consideration by eLife. Your article has been reviewed by three peer reviewers, one of whom, Richard A Neher, is a member of our Board of Reviewing Editors, and the evaluation has been overseen by a Senior Editor. The following individuals involved in review of your submission have agreed to reveal their identity: Katia Koelle (Reviewer #2); Christopher J R Illingworth (Reviewer #3).

Our decision has been reached after consultation between the reviewers. Based on these discussions and the individual reviews below, we regret to inform you that this manuscript will not be considered further for publication in eLife. As detailed below, the reviewers were not convinced by the estimation of within-host Ne but unanimously felt the characterization of the transmission bottleneck is interesting and robust. We, therefore, encourage you to resubmit a manuscript focusing on the transmission bottleneck as a short report.

All reviewers agreed that you have presented comprehensive data on within-host diversity of influenza virus. The data set includes transmission pairs within households which allowed you to estimate the size of the transmission bottleneck. In sharp contrast to previous estimates, your data suggest a very small transmission bottleneck (evident in Figure 3C). This observation is important, and this conclusion is strongly supported by the data. The estimation of Ne, however, is unconvincing. The major concerns regarding the Ne estimation are summarized below -- you can find details in the attached reviews.

1) The signal to estimate Ne is weak. The distribution of iSNV frequency changes as day 3 is hardly different from that at day 1. Furthermore, the distribution of variant frequency changes is long-tailed, which conflicts the assumptions of the Wright-Fisher model. While you correct for false negatives and allow for a constant variance in observed frequency, estimates might still be affected by extraction and amplification noise (particularly important for the second time point with typically lower viral load). At best, your estimate of Ne is a loose lower bound.

2) Whether Ne is a well-defined quantity of an acute infection is debatable. The populations you sequenced may be a mixture of different subpopulations from different patches of infected tissue. Changes in variant frequencies of common iSNVs are therefore might not due to resampling noise, but differential growth and shrinkage of subpopulations. This type of noise is not characterized by an effective population size.

3) Your estimate of the mutation rate (3-4e-5/site/replication) is inconsistent with the 5fold higher rate you estimated in Pauly et al., 2017, despite generous assumptions. How can this difference be reconciled?

Reviewer #1:

McCrone et al., study genetic variation of influenza virus population within hosts and in transmission chain to estimate the stochasticity of within host evolution and the effective size of the transmission bottle neck. Virus populations are characterized by deep sequencing. The quality of the data and the analysis seem high and the conclusions seem justified and solid. I have a number of comments regarding the interpretation of the results and some parts of the analysis.

While the transmission bottleneck size is a relevant and interpretable quantity, I don't think the concept of an effective population is useful for an acute within-host population. The actual population size is orders of magnitudes higher than any estimate of Ne and the variation in variant frequencies likely depends on subpopulations dynamics in the infected tissue. These are populations subject to exponential expansion and decay and there is no reason why a diffusion-based model is useful. This is evident in Figure 2B: The distribution of allele frequency changes is incompatible with a diffusion model: most alleles don't change much in frequency, while others change by 0.4. The removal of the most extremely changing iSNVs partly addresses this issue, but also shows that there is a substantial dependence on this cut-off. The distribution of variant frequency changes is interesting in its own right, but there is probably not enough data to investigate this here. But even taking the estimate of Ne at face value, how is sampling noise handled in the Ne estimates? And why is the 0.02 cutoff a good idea for the later sample? Once a iSNV is ascertained in the earlier sample, you should use the raw variant frequency in the diffusion model.

While fitting a diffusion model is not appropriate, one could still estimate how a quantity like the inter quartile range of (x_1-x_0)/(x_0(1-x_0)) increases with time. A quick and dirty analysis using the data the authors uploaded as supporting information suggests that there is little signal.

The estimate of selection coefficients is not terribly convincing. It seems much more likely that from one day to another, different patches of infected tissue contribute to different degree to the sample and such population shifts will result in changing frequencies.

The estimates of the size of transmission bottleneck, on the other hand, are much better defined and a more relevant quantity than the within host Ne. The estimate seems solid and provide an interesting counterpoint to high estimates from previous studies. It would be interesting to investigate whether child-child, parent-child, and adult-adult transmission have distinct properties.

I find the statements that the potential for spread of a mutation is determined by the transmission bottleneck are misleading. Mutations that rise in frequency within a host have a much higher frequency to be transmitted than other mutations, even if they rise only to frequencies of 1/1000 or less.

To assess the reliability of the variant calling and diversity representation, I would like to see the analogs of Figure 2—figure supplement 1A for samples that were processed in duplicate.

A more systematic investigation of whether antigenic sites tend to segregate within individuals seems possible and important. Why not compare distributions of variant frequencies at sites with global variation in the respective season with those that are globally monomorphic? Similar comparisons could be done for epitope vs non-epitope, or syn vs non-syn.

Any idea why the probability to transmit a 50% variant is larger than 50%? Is this the effect of minor variants and bottlenecks>1 or selective advantage of minor variants? The explanation of the shaded areas is confusing: These are distributions of the inferred probability across 1000 simulations, but not outcomes of transmission.

The method to estimate the mutation rate is problematic. First of all, Eq. 23 is incorrect as written and the model assumptions are unlikely to hold in an infected individual.

Reviewer #2:

In this manuscript, McCrone and coauthors use deep sequencing data from a prospective community-based cohort study to estimate the transmission bottleneck size for seasonal influenza viruses H1N1 and H3N2, as well as the effective population size of the viral population within acutely infected individuals. In contrast to the only other dataset used in quantifying transmission bottleneck sizes for influenza circulating in natural human populations (Poon et al.), they find evidence for a remarkably small bottleneck size of 1-2 influenza virions using two distinct methods. Further, they estimate small within-host effective population sizes of 30-40 virions using a number of different methods. With both small transmission bottleneck sizes and small within-host population sizes, they conclude that genetic drift and stochastic processes are important factors influencing influenza virus dynamics. Finally, they do find some evidence for purifying selection going on within infected hosts, given observed patterns of nonsynonymous versus synonymous nucleotide variation.

In general, this manuscript is very clearly written, is very thorough in applying different methods to arrive at robust conclusions and presents interesting results. One concern is of course the disparity between the results this manuscript presents for transmission bottleneck sizes in flu (1-2 virions) relative to the previous literature estimate of 100-200 virions, based on the data presented in Poon et al. It is clear from the variant frequency plot shown in Figure 3C that these differences in estimated bottleneck sizes are based on differences present in the data themselves, rather than the specifics of the methods applied to the data.

I have several concerns, but none that are sufficiently major to keep this paper from being considered for publication in eLife.

If the within-host effective population size is very small (~35 virions), then it seems to me that the timing of transmission should matter for determining the transmission bottleneck size. Have you looked to see how much the estimates differ from one another based on whether the first or second sampling timepoint in a donor was used? (The data might not be available to conduct this analysis…)

Subsection “Within host populations have low genetic diversity”: I really like that the within-host Ne calculation was done a number of different ways and that similar results were obtained when some of the more stringent assumptions (site independence and large population size) of the first approach were relaxed. One other assumption in all of these models, however, I think is that the viral population size is constant. Given that this is an acute infection, the viral population size is much more likely to be exponentially declining (given that individuals were sampled following symptom onset). How does an exponentially declining viral population affect your estimate of within-host Ne?

Subsection “The mutation rate of influenza A virus within human hosts”: here, you estimate a mutation rate on the order of 3-4 x 10^-5 mutations per site per replication cycle, which corresponds to ~0.35 mutations/genome/replication cycle. You also mention that this is close to your own recently published estimate (Pauly et al., 2017). However, that paper estimated a mutation rate on the order of 1.8 x 10^-4 (H1N1) and 2.5 x 10^-4, resulting in 2-3 mutations/genome per replication cycle. Since that Pauly et al., paper's primary point was to revise the 2.7 x 10^-6 – 3.0 x 10^-5 previous estimates, it would be nice to explicitly state the discrepancy of the Pauly et al., estimate with the estimate obtained in this paper (since they differ by an order of magnitude).

Reviewer #3:

This work presents an interesting dataset collected from a household study of human influenza infection. Via a mathematical analysis, two key claims about the data are presented here. Firstly, the within-host effective population size, Ne, is relatively low, at 30-70. Secondly the transmission bottleneck is tight, with 1-2 viruses typically founding an infection.

There is an odd repetition to the manuscript, in so far as multiple techniques are applied to get each value of N. In so far as a method of analysis is correct, recalculation of the same sum using an alternative approach is unnecessary. In so far as a method has limitations, repeating a calculation using another method, which shares the same limitations, adds no further useful information to the manuscript. Quantitative accuracy is not achieved via a democratic vote among statistical methods.

The analysis of the transmission bottleneck is convincing, though susceptible to criticism under the point above. Given that the presence-absence model underestimates the size of a bottleneck, and the conclusion that the bottleneck is small, why not just use the β-binomial model? For myself, under the assumption that the identification of transmission pairs is correct, Figure 3C is clear evidence of a tight bottleneck.

The analysis of within-host effective population size is problematic. The initial analysis of Ne assumes that selection does not act upon the viral populations, calculating the parameter on the basis of a model of genetic drift alone. This is fine in establishing a lower bound for Ne but cannot support a conclusion that Ne is small. Here, while much care is taken in evaluating both diffusion and Wright-Fisher models, and evaluating the potential power of the calculation, the issues addressed are relatively trifling in the context of the assumption of neutrality. The final result is to conclude that Ne is small, and therefore selection has little influence, all under the assumption that selection can be neglected.

An attempt to fix this problem is made via the use of a method which estimates Ne along with a selection coefficient for each SNP. I am not convinced that this method is effective in the context of the data collected. Whereas the original application of this method was to a dataset for which sequence data was collected at 13 separate time-points (Foll et al., 2014), here only two time-points are available for each patient (Results section). In estimating the value of Ne from time-resolved data, the value of selection, speaking loosely, is estimated from the increase or decrease of an allele frequency over time, while the extent of drift (or Ne) is estimated from the extent of deviation of the data from a deterministic model of selection (see e.g. Feder et al., 2014). Where data is collected for only two time-points, a deterministic model (with effectively infinite Ne) can be fitted perfectly to any allele frequency data, with a different selection coefficient being fitted to each variant. I am therefore unclear where the value of Ne estimated from this method arises from; perhaps fitting a prior to the selection coefficient affects this? I note that the method, while validated for the inference of selection from two time-points, is not validated in the original publication for its ability to infer Ne.

There are other factors which might limit the correct inference of a high Ne. For example, while the existence of variant allele frequencies is well-validated, the precision with which an allele frequency can be measured is less clear. A standard deviation in a variant frequency is cited from a previous paper (subsection “Discrete Wright-Fisher estimation of 𝑁𝑒”), albeit this was measured from in vitro material. Other authors (Lakdawala et al., 2015) have highlighted the potential for non-trivial population structure within a host. Here, while one sample was collected via nasal swab, the second was collected from a mix of nasal swab and throat samples; this could conceivably introduce a greater variance into the allele frequencies. Further, if a constant standard deviation is measured from low-frequency variant calls, this might underestimate the variance in higher-frequency calls, which under a binomial model would be frequency-dependent.

In short, while there are excellent parts of this paper, there are also significant problems. Although I am happy with the lower bound inferred for within-host Ne, I am not convinced what can be said about the upper bound for this parameter. The conclusion of a low within-host effective population size may not be valid.

[Editors’ note: what now follows is the decision letter after the authors submitted for further consideration.]

Thank you for submitting your article "Stochastic processes dominate the within and between host evolution of influenza virus" for consideration by eLife. Your article has been reviewed by three peer reviewers, one of whom, Richard A Neher, is a member of our Board of Reviewing Editors and the evaluation has been overseen by Arup Chakraborty as the Senior Editor. The following individuals involved in review of your submission have agreed to reveal their identity: Katia Koelle (Reviewer #2); Christopher J R Illingworth (Reviewer #3).

The reviewers have discussed the reviews with one another and the Reviewing Editor has drafted this decision to help you prepare a revised submission.

Summary:

McCrone et al., present influenza A virus whole genome deep sequencing data from 200 infected individuals, including 49 likely transmission pairs. The authors find little within-host genetic diversity. Furthermore, low overlap in genetic diversity between samples in a transmission pair suggests that new infections are typically dominated by one founder virus, i.e., the transmission bottleneck is very small. Since the size of the transmission bottleneck is an important determinant of influenza virus transmission and previous studies estimated a much larger bottleneck, the results reported here are an important contribution.

Essential revisions:

This paper is a resubmission of a previously rejected version and was reviewed by the same reviewers. All reviewers agreed (as before) that the results on the transmission bottleneck are important, but two major concerns remain.

1) Limited power to characterize within-host selection. Since most infections seem to be dominated by a single founder virus, the power to detect within-host selection during a few days of observation is very low. Selection would need to amplify a novel beneficial variant (produced at rates of about 1e-5 per day) to above 1% within 3-4 days in order to be detected in this study. Given the difficulty to call variants below 2%, even strong positive selection (s~10%) remains undetectable. Despite this lack of power, the authors make strong claims about the absence of positive selection and often make semi-quantitative statements without a proper comparison to an expectation.

a) Subsection “Mutations in antigenic epitopes are rare”. While you claim stochastic processes dominate within host and positive selection is inefficient, we don't think you have sufficient evidence for this statement. What you could conclude is "positive selection rarely amplifies a beneficial de novo variant above the limit of detection at 2%".

b) In subsection “Within-host populations have low genetic diversity”, you are comparing R^2 values of variant frequencies of same day samples with the fraction of minor variants recovered in later samples. But an R^2 value and the fraction recovered iSNV are not comparable. To make a convincing case that your sequencing method accurately measures iSNV frequency and that the observed frequency changes are biological we would like to see the equivalent of Figure 2—figure supplement 1A for duplicate samples, ideally stratified by viral load (in subsection “Illumina library preparation and sequencing”, you write that samples were processed in duplicate). Such a control is particularly important for the comparison of iSNV frequencies in longitudinal samples, as the samples that are being compared likely differ in viral load. Figure 2—figure supplement 1A could use a 1:1 line. It seems that iSNV frequencies are systematically lower in-home isolates than in clinic isolates for low-frequency variants.

c) Given the difficulty to characterize within host stochasticity and the absence of positive selection, the claims made in this direction need to be revisited, toned down or removed. Furthermore, we suggest changing the title. A title like "Small transmission bottlenecks and low genetic diversity in acute influenza A virus infection". We are open to suggestions here, but please make sure that the title reflects what can be concluded from your data. On larger spatio-temporal scales, influenza virus evolution is certainly not stochastic and the current title doesn't make this distinction clearly enough.

2) Confidence intervals on the transmission bottleneck size. You estimate the parameter of a Poisson distribution that is assumed to describe the distribution of bottlenecks in the population. The error bars of this estimate are remarkably tight given the small number of informative iSNVs in individual pairs. The actual distribution of bottleneck sizes in the population is certainly much broader and fitting a Poisson distribution will underestimate the confidence interval. Please include the 95% confidence intervals for the estimate of each transmission pair in Supplementary file 3. Aggregating these confidence intervals or fitting a broader distribution (say log-normal or power-law) should result in more sensible confidence intervals.

3) Dependence of bottleneck estimates on sample pairings. We suggest, to estimate, bottlenecks from all possible different sample-pairings of donor and recipient in those transmission pairs where either has more than one sample available. In particular the difference between self-sampling and sampling in the clinics (Figure 2—figure supplement 1A suggests a possible systematic deviation) should be tested and discussed.

We suggest resubmitting this manuscript as a short report focusing on transmission bottlenecks. Figure 4 is a plausible supplement to Figure 3.

https://doi.org/10.7554/eLife.35962.035

Author response

[Editors’ note: the author responses to the first round of peer review follow.]

Reviewer #1:

McCrone et al., study genetic variation of influenza virus population within hosts and in transmission chain to estimate the stochasticity of within host evolution and the effective size of the transmission bottle neck. Virus populations are characterized by deep sequencing. The quality of the data and the analysis seem high and the conclusions seem justified and solid. I have a number of comments regarding the interpretation of the results and some parts of the analysis.

While the transmission bottleneck size is a relevant and interpretable quantity, I don't think the concept of an effective population is useful for an acute within-host population. The actual population size is orders of magnitudes higher than any estimate of Ne and the variation in variant frequencies likely depends on subpopulations dynamics in the infected tissue. These are populations subject to exponential expansion and decay and there is no reason why a diffusion-based model is useful. This is evident in Figure 2B: The distribution of allele frequency changes is incompatible with a diffusion model: most alleles don't change much in frequency, while others change by 0.4. The removal of the most extremely changing iSNVs partly addresses this issue, but also shows that there is a substantial dependence on this cut-off. The distribution of variant frequency changes is interesting in its own right, but there is probably not enough data to investigate this here. But even taking the estimate of Ne at face value, how is sampling noise handled in the Ne estimates? And why is the 0.02 cutoff a good idea for the later sample? Once a iSNV is ascertained in the earlier sample, you should use the raw variant frequency in the diffusion model.

While fitting a diffusion model is not appropriate, one could still estimate how a quantity like the inter quartile range of (x_1-x_0)/(x_0(1-x_0)) increases with time. A quick and dirty analysis using the data the authors uploaded as supporting information suggests that there is little signal.

We have removed the diffusion models and all quantitative analysis of effective population size. We have kept the within host data (which the reviewers agreed was of high quality) and discuss it in a qualitative or semi-quantitative manner. We now describe our data as consistent with a model in which variants more frequently arise within hosts through neutral processes such as genetic drift.

The estimate of selection coefficients is not terribly convincing. It seems much more likely that from one day to another different patches of infected tissue contribute to different degree to the sample and such population shifts will result in changing frequencies.

The issue of sampling in a community-based cohort with an acute respiratory viral infection is indeed a challenging one. We note that we provided data in Figure 2—figure supplement 1A of the original manuscript that showed reproducibility of sequence data from two distinct same day samples (one self-collected at home and one collected at the clinic) from six individuals. The reviewer is correct that this does not completely exclude the potential issues of sampling. We have therefore removed the figure and discussion of individual selection coefficients.

The estimates of the size of transmission bottleneck, on the other hand, are much better defined and a more relevant quantity than the within host Ne. The estimate seems solid and provide an interesting counterpoint to high estimates from previous studies. It would be interesting to investigate whether child-child, parent-child, and adult-adult transmission have distinct properties.

We have now provided a Supplementary file 3 that shows the ages and dates of sampling of donor and recipient, bottleneck size, viral subtype, and vaccination status for each transmission pair. We have also generated a shiny app with a link provided (subsection “Identification of forty-three transmission pairs”) so that interested readers can look at the sequence variants observed in each donor and recipient pair.

I find the statements that the potential for spread of a mutation is determined by the transmission bottleneck are misleading. Mutations that rise in frequency within a host have a much higher frequency to be transmitted than other mutations, even if they rise only to frequencies of 1/1000 or less.

If we understand the comment correctly, the reviewer suggests that mutations that increase in frequency within a host have a competitive advantage during transmission, such that a minority variant could be selectively transmitted. This would seem to assume (i) that variants increase in frequency in a host due to positive selection, (ii) that transmission is a selective process as well, (iii) that the same factors are selected within hosts and between hosts. This is a reasonable model, but not well supported by our data. Nevertheless, we have toned down the statement a bit given the uncertainty, and also given that we may misunderstand the comment as stated here. The Introduction now reads. “Their potential for subsequent spread through host populations is heavily dependent on the size of the transmission bottleneck (Alizon et al., 2011; Zwart and Elena, 2015).”

To assess the reliability of the variant calling and diversity representation, I would like to see the analogs of Figure S6 for samples that were processed in duplicate.

The comment seems to suggest that the data presented in Figure 2—figure supplement 1A and elsewhere were based on only one of two samples processed in duplicate. The data are from serial samples (on the same day separated by time) in which variants were ascertained based on duplicate RT-PCR and sequencing libraries for each of the two samples.

As described in the Materials and methods section of the original submission, “To control for PCR errors in samples with lower input titers, all isolates with titers between 103 and 105 genomes/μl were processed and sequenced in duplicate. Only iSNV that were found in both replicates were included in downstream analysis. The frequency of the variant in the replicate with higher coverage at the iSNV location was assigned as the frequency of the iSNV. Finally, any SNV with a frequency below 2% was discarded.”

Therefore, for each sample (nasal/throat swab in 1ml transport media), we performed a single RNA prep. The RNA genome copy number was determined by RT-qPCR. The vast majority of samples had copy numbers between 103 and 105 copies per microliter and were then processed for duplicate multiplex RTPCR reactions and then duplicate Illumina sequencing libraries. We only called iSNV that were present in both replicates subject to our variant calling criteria. We have documented the sensitivity and specificity of this approach in McCrone and Lauring, 2016, used these same criteria in Debbink et al., 2017, and report our validation again in the present manuscript. In our experience, this is well beyond what is often reported in the literature and should suffice for documentation of reliability in variant calling. All figures were based on variants identified using this approach.

A more systematic investigation of whether antigenic sites tend to segregate within individuals seems possible and important. Why not compare distributions of variant frequencies at sites with global variation in the respective season with those that are globally monomorphic? Similar comparisons could be done for epitope vs non-epitope, or syn vs non-syn.

We have now provided a new Figure 2B that shows the distribution of variant frequencies for antigenic and nonantigenic sites in HA (the antigenic sites being the ones likely to be under strongest positive selection). We have also expanded our Figure 2C, which looks at the few within host variants that are found at a global level.

Any idea why the probability to transmit a 50% variant is larger than 50%? Is this the effect of minor variants and bottlenecks>1 or selective advantage of minor variants? The explanation of the shaded areas is confusing: These are distributions of the inferred probability across 1000 simulations, but not outcomes of transmission.

We think the more likely explanation is that it is the effect of minor variants and bottlenecks greater than one, as the reviewer suggests. We observed no convincing evidence of a selective advantage for minor variants across the dataset.

The method to estimate the mutation rate is problematic. First of all, Eq. 23 is incorrect as written and the model assumptions are unlikely to hold in an infected individual.

This section has been removed.

Reviewer #2:

In this manuscript, McCrone and coauthors use deep sequencing data from a prospective community-based cohort study to estimate the transmission bottleneck size for seasonal influenza viruses H1N1 and H3N2, as well as the effective population size of the viral population within acutely infected individuals. In contrast to the only other dataset used in quantifying transmission bottleneck sizes for influenza circulating in natural human populations (Poon et al.), they find evidence for a remarkably small bottleneck size of 1-2 influenza virions using two distinct methods. Further, they estimate small within-host effective population sizes of 30-40 virions using a number of different methods. With both small transmission bottleneck sizes and small within-host population sizes, they conclude that genetic drift and stochastic processes are important factors influencing influenza virus dynamics. Finally, they do find some evidence for purifying selection going on within infected hosts, given observed patterns of nonsynonymous versus synonymous nucleotide variation.

In general, this manuscript is very clearly written, is very thorough in applying different methods to arrive at robust conclusions and presents interesting results. One concern is of course the disparity between the results this manuscript presents for transmission bottleneck sizes in flu (1-2 virions) relative to the previous literature estimate of 100-200 virions, based on the data presented in Poon et al. It is clear from the variant frequency plot shown in Figure 3C that these differences in estimated bottleneck sizes are based on differences present in the data themselves, rather than the specifics of the methods applied to the data.

I have several concerns, but none that are sufficiently major to keep this paper from being considered for publication in eLife.

If the within-host effective population size is very small (~35 virions), then it seems to me that the timing of transmission should matter for determining the transmission bottleneck size. Have you looked to see how much the estimates differ from one another based on whether the first or second sampling timepoint in a donor was used? (The data might not be available to conduct this analysis…)

For samples with two within host samples (2014-2015 season only), we used the time point closest to the time of transmission, as determined by date of onset of symptoms in the recipient. We assume that using this sample would provide a better representation of the diversity in the donor at the time of transmission. As above, we have provided the requested information as part of a shiny app with a link (subsection “Identification of forty-three transmission pairs”) that allows one to examine the variants for each pair across samples.

Subsection “Within host populations have low genetic diversity”: I really like that the within-host Ne calculation was done a number of different ways and that similar results were obtained when some of the more stringent assumptions (site independence and large population size) of the first approach were relaxed. One other assumption in all of these models, however, I think is that the viral population size is constant. Given that this is an acute infection, the viral population size is much more likely to be exponentially declining (given that individuals were sampled following symptom onset). How does an exponentially declining viral population affect your estimate of within-host Ne?

As above, we have removed the Ne analysis.

Subsection “The mutation rate of influenza A virus within human hosts”: here, you estimate a mutation rate on the order of 3-4 x 10^-5 mutations per site per replication cycle, which corresponds to ~0.35 mutations/genome/replication cycle. You also mention that this is close to your own recently published estimate (Pauly et al., 2017). However, that paper estimated a mutation rate on the order of 1.8 x 10^-4 (H1N1) and 2.5 x 10^-4, resulting in 2-3 mutations/genome per replication cycle. Since that Pauly et al., paper's primary point was to revise the 2.7 x 10^-6 – 3.0 x 10^-5 previous estimates, it would be nice to explicitly state the discrepancy of the Pauly et al., estimate with the estimate obtained in this paper (since they differ by an order of magnitude).

As above, we have removed this joint analysis of Ne and mutation rate.

Reviewer #3:

This work presents an interesting dataset collected from a household study of human influenza infection. Via a mathematical analysis, two key claims about the data are presented here. Firstly, the within-host effective population size, Ne, is relatively low, at 30-70. Secondly the transmission bottleneck is tight, with 1-2 viruses typically founding an infection.

There is an odd repetition to the manuscript, in so far as multiple techniques are applied to get each value of N. In so far as a method of analysis is correct, recalculation of the same sum using an alternative approach is unnecessary. In so far as a method has limitations, repeating a calculation using another method, which shares the same limitations, adds no further useful information to the manuscript. Quantitative accuracy is not achieved via a democratic vote among statistical methods.

The analysis of the transmission bottleneck is convincing, though susceptible to criticism under the point above. Given that the presence-absence model underestimates the size of a bottleneck, and the conclusion that the bottleneck is small, why not just use the β-binomial model? For myself, under the assumption that the identification of transmission pairs is correct, Figure 3C is clear evidence of a tight bottleneck.

There is no consensus on the right way to analyze a bottleneck. We performed both, not because we pursued a “democratic vote,” but as a way to ensure that our estimates are robust to the assumptions of each model. We are aware of issues regarding presence absence vs. β binomial. The presence-absence model actually provided a marginally better fit for our data. We suspect that if we didn’t present the β binomial, a reviewer would ask that we analyze our data under that model (given the time between transmission and sampling).

The analysis of within-host effective population size is problematic. The initial analysis of Ne assumes that selection does not act upon the viral populations, calculating the parameter on the basis of a model of genetic drift alone. This is fine in establishing a lower bound for Ne but cannot support a conclusion that Ne is small. Here, while much care is taken in evaluating both diffusion and Wright-Fisher models, and evaluating the potential power of the calculation, the issues addressed are relatively trifling in the context of the assumption of neutrality. The final result is to conclude that Ne is small, and therefore selection has little influence, all under the assumption that selection can be neglected.

We have removed analyses related to Ne.

An attempt to fix this problem is made via the use of a method which estimates Ne along with a selection coefficient for each SNP. I am not convinced that this method is effective in the context of the data collected. Whereas the original application of this method was to a dataset for which sequence data was collected at 13 separate time-points (Foll et al., 2014), here only two time-points are available for each patient (Results section). In estimating the value of Ne from time-resolved data, the value of selection, speaking loosely, is estimated from the increase or decrease of an allele frequency over time, while the extent of drift (or Ne) is estimated from the extent of deviation of the data from a deterministic model of selection (see e.g. Feder et al., 2014). Where data is collected for only two time-points, a deterministic model (with effectively infinite Ne) can be fitted perfectly to any allele frequency data, with a different selection coefficient being fitted to each variant. I am therefore unclear where the value of Ne estimated from this method arises from; perhaps fitting a prior to the selection coefficient affects this? I note that the method, while validated for the inference of selection from two time-points, is not validated in the original publication for its ability to infer Ne.

We have removed this analysis, which was related to Ne.

There are other factors which might limit the correct inference of a high Ne. For example, while the existence of variant allele frequencies is well-validated, the precision with which an allele frequency can be measured is less clear. A standard deviation in a variant frequency is cited from a previous paper (subsection “Discrete Wright-Fisher estimation of 𝑁𝑒”), albeit this was measured from in vitro material. Other authors (Lakdawala et al., 2015) have highlighted the potential for non-trivial population structure within a host. Here, while one sample was collected via nasal swab, the second was collected from a mix of nasal swab and throat samples; this could conceivably introduce a greater variance into the allele frequencies.

The reviewer is correct, and this is a challenging aspect of studies in natural infections with acute respiratory viruses. The standard deviation in variant frequency is indeed from a benchmarking experiment with in vitro material. However, given that viruses were diluted at set frequencies in media similar to the universal transport media used in clinical specimens and at genome equivalents similar to the range used in the study, we think that this provides a reasonable estimate of error. We struggle to come up with a better experiment that would provide some sort of estimate of precision given the lack of an in vivo “gold standard” for comparison. We acknowledge that there could certainly be spatial stratification as the reviewer suggests. However, as above, we note that Figure 2—figure supplement 1 shows that replicate same-day samples from 6 individuals separated by hours – one nasal and one nasal/throat – were well correlated with respect to iSNV type and frequency. For comparison, the excellent work of Lakdawala et al., cited by the reviewer reports spatial differences in allele frequencies for 3 ferrets who were experimentally infected.

[Editors' note: the author responses to the re-review follow.]

Essential revisions:

This paper is a resubmission of a previously rejected version and was reviewed by the same reviewers. All reviewers agreed (as before) that the results on the transmission bottleneck are important, but two major concerns remain.

1) Limited power to characterize within-host selection. Since most infections seem to be dominated by a single founder virus, the power to detect within-host selection during a few days of observation is very low. Selection would need to amplify a novel beneficial variant (produced at rates of about 1e-5 per day) to above 1% within 3-4 days in order to be detected in this study. Given the difficulty to call variants below 2%, even strong positive selection (s~10%) remains undetectable. Despite this lack of power, the authors make strong claims about the absence of positive selection and often make semi-quantitative statements without a proper comparison to an expectation.

We agree with this critique and recognizing that “absence of evidence is not evidence of absence,” we were careful in our manuscript to indicate that while positive selection is an inefficient driver over the course of an acute infection, it is definitely not absent. We concede that due to well-known issues in variant calling in NGS data, we only really make our arguments based on variants present at greater than 2% frequency and that this imposes a limit on what we can say about positive selection in an acute infection.

In response to the suggestions from these same reviewers from the previously rejected manuscript, we included additional analyses to look for evidence of selection with the suggested null models. First, we compared the number of HA variants in antigenic and non-antigenic sites (the former being more likely to be subject to positive selection) and found that our data were consistent with the null expectation of no difference (Figure 2B). Second, we asked whether any of the antigenic variants identified as minority SNV in our study were subsequently observed globally, as this would support the idea that they were “on the way up” due to positive selection. We found that this did not appear to be the case (Figure 2C). Finally, with the additional data presented now regarding the accuracy of our frequency measurements (see below), we can state more strongly that the data in Figure 2E regarding the dynamics of iSNV within hosts show evidence for considerable stochasticity. Here, the null expectation is that synonymous variants will show similar dynamics as antigenic variants (the latter being most likely to be affected by positive selection). We note that the number of variants detectable at >2% is roughly consistent with a neutral model based on influenza’s known mutation rate and generation time (see for example Box and figures in Xue et al., 2018.

In sum, we have performed several distinct analyses to identify evidence for positive selection in acute influenza infections, each with a reasonable null model. From these, we concluded that while positive selection clearly happens within hosts, it is not sufficiently strong to frequently drive new variants to fixation or even to levels where they are likely to be transmitted (our data would estimate that the probability of transmission is approximately 1.7% for a variant at 1% frequency and 3.3% for a variant at 2% frequency in both presence-absence and β-binomial models). Defining the limits of positive selection vis a vis drift at the level of individual hosts is an important aspect of our work, as three recent reviews on influenza evolution offer differing views on this question. (see Johnson et al., 2017; Petrova and Russell, 2017; Xue et al., 2018).

a) Subsection “Mutations in antigenic epitopes are rare”. While you claim stochastic processes dominate within host and positive selection is inefficient, we don't think you have sufficient evidence for this statement. What you could conclude is "positive selection rarely amplifies a beneficial de novo variant above the limit of detection at 2%".

We agree with this statement and this is largely how we discuss the data. We concede that our power to detect positive selection during acute selection is limited by the frequency of sampling and the reliability of variant detection below 2%. To address these concerns and limitations, we have further clarified our arguments and toned down any arguments regarding positive selection, see the following sentences (which include nearly all references to positive selection in the manuscript).

Introduction “Despite limited evidence for positive selection, novel mutations do arise within hosts and some will clearly be positively selected. Their potential for subsequent spread through host populations is heavily dependent on the size of the transmission bottleneck (Alizon et al., 2011; Zwart and Elena, 2015).”

Introduction “We find that intrahost populations are dynamic and constrained by genetic drift and purifying selection. In our study, positive selection rarely amplifies a beneficial de novo variant to a frequency greater than 2%.”

Subsection “Within host populations are dynamic” “Together with our prior work (Debbink et al., 2017), these data suggest that the positive selection of novel variants within hosts is inefficient and rarely amplifies a newly generated variant to a frequency greater than 2%.”

Subsection “Identification of forty-three transmission pairs” “Our within-host data suggest that newly arising iSNV with positive fitness effects are likely to be present at low frequencies (<2%) during an acute infection.”

Discussion section “Contrary to previous studies, which have found signatures of adaptive evolution in infected hosts (Gubareva et al., 2001; Rogers et al., 2015; Ghedin et al., 2011; Sobel Leonard et al., 2016), we have found only limited evidence of positive selection during acute infection. Previous reports have relied on infections in which selective pressures are likely to be particularly strong (e.g. due to drug treatment or infection with a poorly adapted virus), or in which the virus has replicated within a host for an extended period of time (Xue et al., 2017). Under these conditions, it is plausible that positively selected alleles may reach levels of detection. We suggest that these are important and informative exceptions to dynamics defined here, in which positive selection is rarely strong enough to drive a new mutation to a frequency above 2% over the course of several days.”

Discussion section “We have deeply sequenced 332 intrahost populations from 283 individuals collected over more than 11,000 person-seasons of observation and only identified a handful of minority antigenic variants with limited evidence for positive selection (this work and (Debbink et al., 2017)). Importantly, our data suggest that even if selection acts below our level of detection, such rare variants are unlikely to transmit. Given the size of the estimated bottleneck, the probability of transmission is approximately 1.7% for a variant at 1% frequency and 3.3% for a variant at 2% frequency. However, with several million infected individuals each year, even inefficient processes and rare events at the scale of individual hosts are likely to occur at a reasonable frequency on a global scale.”

b) In subsection “Within-host populations have low genetic diversity”, you are comparing R^2 values of variant frequencies of same day samples with the fraction of minor variants recovered in later samples. But an R^2 value and the fraction recovered iSNV are not comparable. To make a convincing case that your sequencing method accurately measures iSNV frequency and that the observed frequency changes are biological we would like to see the equivalent of Figure 2—figure supplement 1 for duplicate samples, ideally stratified by viral load (in subsection “Illumina library preparation and sequencing”, you write that samples were processed in duplicate). Such a control is particularly important for the comparison of iSNV frequencies in longitudinal samples, as the samples that are being compared likely differ in viral load. Figure 2—figure supplement 1 could use a 1:1 line. It seems that iSNV frequencies are systematically lower in-home isolates than in clinic isolates for low-frequency variants.

We appreciate these suggestions and have provided the requested analyses. We have added a 1:1 line to Figure 2—figure supplement 1A. There are several variants that were identified above the 2% cutoff in one sample but not the other. Together with the missing 1:1 line, this gave the appearance of a systematic bias in frequency measurement. As suggested, we have supplied a new Figure 2—figure supplement 1B, which shows iSNV frequency measurements across all duplicate samples stratified by viral load. Note that samples above 105 copies per microliter were not processed in duplicate and those with fewer than 103 copies per microliter were not used. This figure shows that our sequencing method accurately measures iSNV frequency. Finally, we have supplied a new Figure 2—figure supplement 1C, which compares the measurement error in frequency measurement (from duplicate RT-PCR and sequencing libraries) to the observed frequency differences in the serial biological samples. The impact of measurement error is minor. Together, these data strengthen our case regarding the biological significance of the changes in frequency in serial samples. They establish the bounds of what can reasonably be explained by stochastic and selective forces within hosts.

c) Given the difficulty to characterize within host stochasticity and the absence of positive selection, the claims made in this direction need to be revisited, toned down or removed. Furthermore, we suggest changing the title. A title like "Small transmission bottlenecks and low genetic diversity in acute influenza A virus infection". We are open to suggestions here, but please make sure that the title reflects what can be concluded from your data. On larger spatio-temporal scales, influenza virus evolution is certainly not stochastic and the current title doesn't make this distinction clearly enough.

As above, we have toned down where appropriate. We feel that our controls in variant calling and measurement error, our analyses of antigenic variants, and our quantitative analysis of SNV dynamics in serial samples demonstrate that while present, positive selection is an inefficient driver of variant accumulation within hosts – largely due to the short time scales (as summarized by the editor above). As we wrote in the Introduction of the submitted manuscript, “Despite limited evidence for positive selection, novel mutations do arise within hosts. Their potential for subsequent spread through host populations is heavily dependent on the size of the transmission bottleneck (Alizon et al., 2011; Zwart and Elena, 2015).” Therefore, we believe the within host data are important to the field, as they allow us to put the bottleneck in context, beyond simply providing a number. The data in Figure 2E suggest that variants arising through neutral processes are just as likely to be present at frequencies above 2% and therefore transmitted. Furthermore, the small size of the bottleneck and the good fit of a neutral model of transmission provide strong evidence for stochasticity. An alternative title that might more concisely capture the impact of stochasticity without overstating its role would be, “Stochastic processes constrain the within and between host evolution of influenza virus.” This title also clearly delineates the scale at which these processes are studied (individual hosts). We note that stochasticity in influenza evolution has been documented at larger scales (bigger than hosts, smaller than globally) in a paper by Nelson et al., “Stochastic processes are key determinants of short-term evolution in influenza”, 2006.

2) Confidence intervals on the transmission bottleneck size. You estimate the parameter of a Poisson distribution that is assumed to describe the distribution of bottlenecks in the population. The error bars of this estimate are remarkably tight given the small number of informative iSNVs in individual pairs. The actual distribution of bottleneck sizes in the population is certainly much broader and fitting a Poisson distribution will underestimate the confidence interval. Please include the 95% confidence intervals for the estimate of each transmission pair in Supplementary file 3. Aggregating these confidence intervals or fitting a broader distribution (say log-normal or power-law) should result in more sensible confidence intervals.

We have added the 95% confidence intervals (using the β binomial model) for the estimate of each transmission pair in Supplementary file 3. As expected, they are quite large because each transmission pair has so few data points. We have aggregated these confidence intervals using a log-normal distribution as suggested and weighted each transmission pair by the number of donor iSNV as performed in Sobel Leonard et al., 2017 (the same study in which the β binomial model was first implemented). The associated text now reads, “We also fit a long-tailed, discrete distribution based on the lognormal. As expected, this analysis resulted in a slightly wider distribution with a mode of 1, a 95th percentile of 11 and a 97.5th percentile of 21.” This analysis was run up to a maximum bottleneck of 1000. We have also added a curve to the bottleneck models in Figure 3 showing the probability that a variant will be transmitted given its frequency in the donor and a bottleneck of 10. This gives an indication of the fit of our aggregate bottleneck to the data.

In addressing this comment and those above, we noticed a “bug” in our code for plotting the β-binomial model in Figure 4E (now Figure 3E). The estimates and model fit as reported in the text were correct, but the plot was not. We have replaced this panel with the correct plot and apologize for our error. Consistent with the similarities in the results obtained through the presence-absence and β-binomial models, this plot is quite similar, but not identical to the one in Figure 3D.

3) Dependence of bottleneck estimates on sample pairings. We suggest, to estimate, bottlenecks from all possible different sample-pairings of donor and recipient in those transmission pairs where either has more than one sample available. In particular the difference between self-sampling and sampling in the clinics (Figure 2—figure supplement 1 suggests a possible systematic deviation) should be tested and discussed.

In the submitted manuscript we used the sample (home vs. clinic) that was obtained closer to the estimated time of transmission (based on onset of symptoms in the recipient). If the sample closest to the time of transmission had no minority SNV above 2%, we used the other sample (this is more conservative as we attempted to use all possible SNV data for these pairs). While we do not think that the data in Figure 2—figure supplement 1Ashow a systematic deviation, we do acknowledge that the choice of sample for these paired isolates from the 2014-2015 season entails certain assumptions about the transmission process. We have provided bottleneck estimates for all four possible combinations for which there are paired samples for each member of a pair as a source data file (Figure 3—source data 7).

We suggest resubmitting this manuscript as a short report focusing on transmission bottlenecks. Figure 4 is a plausible supplement to Figure 3.

We have moved Figure 4 to the supplement (now Figure 3—figure supplement 1) as suggested.

https://doi.org/10.7554/eLife.35962.036

Article and author information

Author details

  1. John T McCrone

    Department of Microbiology and Immunology, University of Michigan, Ann Arbor, United States
    Contribution
    Resources, Data curation, Software, Formal analysis, Investigation, Visualization, Methodology, Writing—original draft, Writing—review and editing
    Competing interests
    No competing interests declared
    ORCID icon "This ORCID iD identifies the author of this article:" 0000-0002-9846-8917
  2. Robert J Woods

    Division of Infectious Diseases, Department of Internal Medicine, University of Michigan, Ann Arbor, United States
    Contribution
    Formal analysis, Investigation, Writing—review and editing
    Competing interests
    No competing interests declared
  3. Emily T Martin

    Department of Epidemiology, University of Michigan, Ann Arbor, United States
    Contribution
    Resources, Data curation, Project administration, Writing—review and editing
    Competing interests
    No competing interests declared
  4. Ryan E Malosh

    Department of Epidemiology, University of Michigan, Ann Arbor, United States
    Contribution
    Data curation, Writing—review and editing
    Competing interests
    No competing interests declared
    ORCID icon "This ORCID iD identifies the author of this article:" 0000-0003-3546-5935
  5. Arnold S Monto

    Department of Epidemiology, University of Michigan, Ann Arbor, United States
    Contribution
    Data curation, Investigation, Methodology, Project administration, Writing—review and editing
    Competing interests
    No competing interests declared
  6. Adam S Lauring

    1. Department of Microbiology and Immunology, University of Michigan, Ann Arbor, United States
    2. Division of Infectious Diseases, Department of Internal Medicine, University of Michigan, Ann Arbor, United States
    Contribution
    Conceptualization, Data curation, Formal analysis, Supervision, Funding acquisition, Investigation, Methodology, Writing—original draft, Project administration, Writing—review and editing
    For correspondence
    alauring@med.umich.edu
    Competing interests
    No competing interests declared
    ORCID icon "This ORCID iD identifies the author of this article:" 0000-0003-2906-8335

Funding

National Institute of General Medical Sciences (T32 GM007544)

  • John T McCrone

National Institute of Allergy and Infectious Diseases (K08 AI119182)

  • Robert J Woods

Centers for Disease Control and Prevention (U01 IP00474)

  • Arnold S Monto

National Institute of Allergy and Infectious Diseases (R01 AI097150)

  • Arnold S Monto

Doris Duke Charitable Foundation (CSDA 2013105)

  • Adam S Lauring

National Institute of Allergy and Infectious Diseases (R01 AI118886)

  • Adam S Lauring

University of Michigan (Discovery grant)

  • Adam S Lauring

The funders had no role in study design, data collection and interpretation, or the decision to submit the work for publication.

Acknowledgements

This work was supported by a Clinician Scientist Development Award from the Doris Duke Charitable Foundation (CSDA 2013105), a University of Michigan Discovery Grant, and R01 AI118886, all to ASL. The HIVE cohort was supported by NIH R01 AI097150 and CDC U01 IP00474 to ASM. JTM was supported by the Michigan Predoctoral Training Program in Genetics (T32GM007544). RJW was supported by K08 AI119182. We thank Katherine Xue for suggestions on analysis of globally circulating variants and Alexey Kondrashov and Aaron King for helpful discussion.

Ethics

Human subjects: This study was approved by the Institutional Review Board of the University of Michigan Medical School. Adults provided written informed consent for participation for themselves and their children; children 7-17 years provided oral assent.

Reviewing Editor

  1. Richard A Neher, University of Basel, Switzerland

Publication history

  1. Received: February 15, 2018
  2. Accepted: April 18, 2018
  3. Accepted Manuscript published: April 23, 2018 (version 1)
  4. Version of Record published: May 3, 2018 (version 2)
  5. Version of Record updated: June 29, 2018 (version 3)

Copyright

© 2018, McCrone et al.

This article is distributed under the terms of the Creative Commons Attribution License, which permits unrestricted use and redistribution provided that the original author and source are credited.

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