1. Epidemiology and Global Health
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Childhood injury after a parental cancer diagnosis

  1. Ruoqing Chen  Is a corresponding author
  2. Amanda Regodón Wallin
  3. Arvid Sjölander
  4. Unnur Valdimarsdóttir
  5. Weimin Ye
  6. Henning Tiemeier
  7. Katja Fall
  8. Catarina Almqvist
  9. Kamila Czene
  10. Fang Fang
  1. Karolinska Institutet, Sweden
  2. University of Iceland, Iceland
  3. Erasmus MC University Medical Center, The Netherlands
  4. Örebro University, Sweden
  5. Karolinska University Hospital, Sweden
Research Article
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Cite this article as: eLife 2015;4:e08500 doi: 10.7554/eLife.08500

Abstract

A parental cancer diagnosis is psychologically straining for the whole family. We investigated whether a parental cancer diagnosis is associated with a higher-than-expected risk of injury among children by using a Swedish nationwide register-based cohort study. Compared to children without parental cancer, children with parental cancer had a higher rate of hospital contact for injury during the first year after parental cancer diagnosis (hazard ratio [HR] = 1.27, 95% confidence interval [CI] = 1.22-1.33), especially when the parent had a comorbid psychiatric disorder after cancer diagnosis (HR = 1.41, 95% CI = 1.08-1.85). The rate increment declined during the second and third year after parental cancer diagnosis (HR = 1.10, 95% CI = 1.07-1.14) and became null afterwards (HR = 1.01, 95% CI = 0.99-1.03). Children with parental cancer also had a higher rate of repeated injuries than the other children (HR = 1.13, 95% CI = 1.12-1.15). Given the high rate of injury among children in the general population, our findings may have important public health implications.

https://doi.org/10.7554/eLife.08500.001

eLife digest

A diagnosis of cancer can be devastating for both a person and his or her family. Over the past 40 years, the number of individuals in Sweden diagnosed with cancer has more than doubled leaving growing numbers of families coping with the aftermath. Many individuals diagnosed with cancer have young children. Parents with cancer and their spouses often struggle to cope with disease and the demands of parenting simultaneously. In fact, previous research has shown children with a parent who has cancer have a greater risk of behavioral problems or distress than children with two healthy parents.

Whether the stress of having a parent with cancer also affects the children’s physical wellbeing hasn’t been studied much. One concern in particular is whether these children may be at increased risk of injury. Injuries are the most common reason for a child to visit a hospital and in some cases lead to deaths. Children who are not well supervised or whose parents have poor mental health are at increased risk of injury. Coping with cancer and the mental anguish it causes may distract parents and possibly place their children at increased risk of injury.

Based on data from nationwide population and health registers in Sweden, Chen, Regodón Wallin et al. now provide evidence that a child with a parent who has cancer is at a greater risk of injury than a child with two parents who are free of cancer. The analysis also revealed that the risk is particularly great if the parent with cancer also develops mental illness after the cancer diagnosis. The risk of injury is greatest in the first year after the parent’s diagnosis. Fortunately, the elevated risk of injury decreases overtime and is almost non-existing after the third year. The analyses suggest that providing extra support for parents with cancer might help to reduce the risk of injury in their children.

https://doi.org/10.7554/eLife.08500.002

Introduction

Cancer is not only a leading cause of morbidity and mortality among the affected patients, it is also increasingly recognized as a contributor to ill-health in their significant others (Sjovall et al., 2009; Visser et al., 2004; Kazak et al., 2005; Krauel et al., 2012). In Sweden, the number of newly diagnosed cancer patients has more than doubled during the last forty years and a considerable proportion of these patients are parenting minor children (National Board of Health and Welfare, 2014). A diagnosis of cancer in parents has repeatedly been shown to exert psychological and social stress in their children (Visser et al., 2004). Coping with cancer may affect the parenting of both the ill and well parents, further impacting the behavioral and social adaptability of the children (Faulkner and Davey, 2002). In contrast to the relatively rich literature on behavioral and mental well-being of children living with a parent with cancer, few studies have so far addressed somatic health outcomes among these children. In a recent study, we reported that children of parents with cancer had a higher risk of death, both due to cancer and other causes (Chen et al., 2015).

Injury is the most common cause of hospital care among children and accounts for almost one million child deaths annually worldwide (Peden et al., 2008). Sociodemographic, behavioral and psychosocial factors of both children and their family are known determinants of injuries among children (Horwitz et al., 1988). For example, childhood injury has been associated with male sex, risk-taking behavior, lack of parental supervision as well as poor mental health of the parents (Matheny, 1986; Schwebel et al., 2011; Morrongiello et al., 2006b; Peden et al., 2008; McKinlay et al., 2010). To our knowledge, no study has however specifically addressed the impact of parental cancer diagnosis on the risk of child injury. To this end, we leveraged the nationwide population and health registers in Sweden to explore the association between parental cancer diagnosis and the risk of hospital contact for injury among children.

Results

During the study period, 78,815 children (4%) were exposed to a parental cancer diagnosis. The general characteristics of the exposed children differed slightly from the unexposed children, in terms of larger number of siblings, shorter gestational age, higher proportion of delivery through caesarian section, higher proportions of birth weight <2500 or >4000 grams, higher proportion of maternal smoking in early pregnancy, higher paternal and maternal ages at child's birth, as well as higher educational level of the parents (Table 1).

Table 1

Characteristics of the participating children and their parents.

https://doi.org/10.7554/eLife.08500.003
All childrenChildren with
parental cancer
Children without
parental cancer
CharacteristicsN (%)N (%)N (%)p
Characteristics of the children
Sex
 Male1,008,982 (51.4)40,527 (51.4)968,455 (51.4)0.72
 Female955,645 (48.6)38,288 (48.6)917,357 (48.6)
No. of siblings and half siblings
 0189,556 (9.6)5,754 (7.3)

183,802 (9.7)

 1784,911 (40.0)28,901 (36.7)756,010 (40.1)<0.001
 2564,343 (28.7)23,291 (29.6)541,052 (28.7)
 ≥ 3425,817 (21.7)20,869 (26.5)404,948 (21.5)
Gestational age (weeks)
 < 3542,376 (2.2)1,833 (2.3)40,543 (2.1)
 35 - 3670,999 (3.6)3,111 (3.9)67,888 (3.6)
 37 - 38363,508 (18.5)15,192 (19.3)348,316 (18.5)
 39 - 40973,949 (49.6)38,169 (48.4)935,780 (49.6)<0.001
 41 - 42456,222 (23.2)18,333 (23.3)437,889 (23.2)
 ≥ 4312,898 (0.7)528 (0.7)12,370 (0.7)
 Missing44,675 (2.3)1,649 (2.1)43,026 (2.3)
Mode of delivery
 Caesarean section237,822 (12.1)10,300 (13.1)227,522 (12.1)
 Vaginal delivery1,684,729 (85.8)66,971 (85.0)1,617,758 (85.8)<0.001
 Missing42,076 (2.1)1,544 (2.0)40,532 (2.1)
Birth weight (g)
 < 250078,412 (4.0)3,455 (4.4)74,957 (4.0)
 2500-2999207,951 (10.6)8,361 (10.6)199,590 (10.6)
 3000-3499604,528 (30.8)23,466 (29.8)581,062 (30.8)
 3500-3999665,343 (33.9)26,714 (33.9)638,629 (33.9)<0.001
 4000-4499290,798 (14.8)12,039 (15.3)278,759 (14.8)
 ≥ 450068,620 (3.5)2,980 (3.8)65,640 (3.5)
 Missing48,975 (2.5)1,800 (2.3)47,175 (2.5)
Maternal smoking in early pregnancy
 No1,421,392 (72.4)55,388 (70.3)

1,366,004 (72.4)

 Yes383,760 (19.5)17,036 (21.6)366,724 (19.5)<0.001
 Missing159,475 (8.1)6,391 (8.1)153,084 (8.1)
Characteristics of the parents
Paternal age at child's birth (years)
 < 2011,942 (0.6)

 166 (0.2)

11,776 (0.6)
 20-24199,251 (10.1)

4,232 (5.4)

195,019 (10.3)
 25-29584,302 (29.7)16,318 (20.7)567,984 (30.1)<0.001
 30-34628,352 (32.0)23,309 (29.6)605,043 (32.1)
 ≥ 35540,780 (27.5)34,790 (44.1)505,990 (26.8)
Maternal age at child's birth (years)
 < 2047,255 (2.4)840 (1.1)46,415 (2.5)
 20-24395,011 (20.1)9,283 (11.8)385,728 (20.5)
 25-29713,827 (36.3)23,318 (29.6)690,509 (36.6)<0.001
 30-34547,990 (27.9)25,787 (32.7)522,203 (27.7)
 ≥ 35260,544 (13.3)19,587 (24.9)240,957 (12.8)
Highest educational level
 Primary school or lower98,230 (5.0)4,144 (5.3)94,086 (5.0)
 Secondary education987,431 (50.3)36,105 (45.8)951,326 (50.5)
 Tertiary education844,523 (43.0)36,812 (46.7)807,711 (42.8)<0.001
 Postgraduate education34,063 (1.7)1,748 (2.2)32,315 (1.7)
 Missing380 (0.0)6 (0.0)374 (0.0)

Primary analysis

During follow-up, 15,377 exposed children (incidence rate: 52 per 1000 person-years) and 548,488 unexposed children (incidence rate: 46 per 1000 person-years) had a first hospital contact for injury. Adjusting for only attained age and sex, the exposed children had a 4% higher rate of hospital contact for injury (hazard ratio [HR], 1.04; 95% confidence interval [CI], 1.02–1.05) than the other children. After adjustment for all covariates, the association became stronger (HR, 1.07 [95% CI 1.05–1.09]) (Table 2). Approximately 17% of hospital contacts among the exposed children occurred during the first year after cancer diagnosis, corresponding to an incidence rate of 60 per 1000 person-years and a HR of 1.27 (95% CI 1.22–1.33). The rate increment decreased during the second and third years, and became null after three years (Table 2).

Table 2

Hazard ratios for hospital contact for injury among children with parental cancer compared to children without parental cancer.

https://doi.org/10.7554/eLife.08500.004
Any Time After Parental Cancer DiagnosisFirst Year After Parental Cancer Diagnosis
CharacteristicsNo. of Children With a Hospital Contact for InjuryPerson-yearsHR (95%CI) *p (Wald Test)No. of Children With a Hospital Contact for InjuryPerson-yearsHR (95%CI) *p (Wald Test)
No parental cancer548,48811,879,0751548,48811,879,0751
Parental cancer15,377298,3021.07 (1.05-1.09)2,67444,6001.27 (1.22-1.33)
 Time since cancer diagnosis
 ≤ 1 year2,67444,6001.27 (1.22-1.33)
 >1 and ≤3 years3,85074,0871.10 (1.07-1.14)<0.001
 > 3 years8,853179,6151.01 (0.99-1.03)
 Sex of the cancer parent
 Male6,554126,2771.08 (1.05-1.11)0.481,16618,9171.32 (1.24-1.40)0.13
 Female8,823172,0261.06 (1.04-1.09)1,50825,6831.24 (1.18-1.31)
 Tobacco-related cancer
 No12,008233,8481.07 (1.05-1.09)0.722,14235,0801.29 (1.24-1.35)0.13
 Yes3,36964,4541.08 (1.04-1.12)5329,5201.20 (1.10-1.31)
 Alcohol-related cancer
 No10,464201,3891.08 (1.05-1.10)0.301,74528,5251.30 (1.24-1.37)0.16
 Yes4,91396,9131.06 (1.02-1.09)92916,0761.23 (1.15-1.31)
 Predicted 5-year relative survival rate
 < 20% §93118,8451.02 (0.95-1.10)1603,0411.15 (0.98-1.35)
 20-80%7,112136,0801.08 (1.06-1.11)0.211,24320,7361.27 (1.19-1.35)0.38
 ≥ 80% 7,334143,3771.06 (1.04-1.09)1,27120,8241.30 (1.23-1.38)
 Parental psychiatric comorbidity after cancer diagnosis
 No14,630285,6211.06 (1.05-1.08)0.0012,61143,6631.27 (1.22-1.32)0.45
 Yes74712,6811.21 (1.12-1.31)639381.41 (1.08-1.85)
  1. HR, hazard ratio; CI, confidence interval

  2. *Adjusted for attained age, sex, number of siblings, gestational age, mode of delivery and birth weight of the child, paternal age at child's birth, maternal age at child's birth, maternal smoking during early pregnancy, and the highest educational level of the parents.

  3. Tobacco-related cancers include cancers in lung, oesophagus, larynx, pharynx, mouth, lip, salivary glands, tongue, stomach, urinary bladder, kidney, uterine cervix, colon and pancreas.

  4. Alcohol-related cancers include cancers in liver, oral cavity, pharynx, larynx, oesophagus, colorectum and breast.

  5. §Including cancers in esophagus, liver, gall bladder, biliary tract, pancreas, lung and stomach.

  6. Including cancers in lip, breast, corpus uteri, testis, skin, thyroid and other endocrine glands, and Hodgkin’s lymphoma.

  7. Including depression, anxiety disorders, stress reaction and adjustment disorder.

The association was not modified by the sex of cancer parent or by the expected survival of cancer; the association did not differ between smoking/alcohol-related cancers and other cancers (Table 2; all p>0.05). However, children whose cancer parent had developed a comorbid psychiatric disorder after diagnosis had a higher rate of childhood injury (HR, 1.21 [95% CI 1.12–1.31], compared with children whose cancer parent had no such disease (HR, 1.06 [95% CI 1.05–1.08]) (p = 0.001). As in the overall analysis, the rate increment in these analyses was more prominent during the first year after diagnosis (Table 2).

The overall association was significantly stronger for boys than for girls (p for interaction, < 0.001) (Table 3). When focusing on the first year following parental cancer, no statistically significant difference was however detected between boys and girls (p = 0.17). Neither the overall association nor the association during the first year after parental cancer was modified otherwise by age at follow-up or number of siblings of the child (Table 3).

Table 3

Hazard ratios for hospital contact for injury among children with parental cancer compared to children without parental cancer, according to sex, age and number of full and half siblings of the child.

https://doi.org/10.7554/eLife.08500.005
No Parental CancerAny Time After Parental Cancer DiagnosisFirst Year After Parental Cancer Diagnosis
Characteristics of the ChildNo. of Children With a Hospital Contact for InjuryPerson-yearsHR (95%CI)No. of Children With a Hospital Contact for InjuryPerson-yearsHR(95%CI)p for interactionNo. of Children With a Hospital Contact for InjuryPerson-yearsHR (95%CI)p for interaction
Sex*
 Male313,8065,966,451

1

9,088150,070
1.11
(1.08-1.13)
< 0.0011,594

22,747

1.30
(1.24-1.37)
0.17
 Female234,6825,912,62416,289148,2331.02
(0.99-1.05)
1,08021,8541.23
(1.16-1.31)
Age (years)
 < 335,157876,76111032,1061.21
(0.99-1.47)
571,1551.25
(0.96-1.63)
 3-555,4521,508,188143911,7751.07
(0.97-1.18)
1313,1811.19
(1.00-1.42)§
 6-11197,9844,625,78614,08588,1061.08
(1.05-1.12)
0.6075614,4351.24
(1.15-1.34)
0.72
 12-15134,9952,500,21614,69883,1991.08
(1.04-1.11)
77711,4291.27
(1.18-1.37)
 ≥ 15124,9002,368,12416,052113,1161.06
(1.03-1.09)
95314,4001.32
(1.23-1.41)
No. of full and half siblings
 056,1741,346,43511,14324,3011.05
(0.98-1.11)
208

3,759

1.30
(1.13-1.50)
 1229,1454,982,31015,806113,7721.06
(1.04-1.09)
0.1398716,9071.24
(1.16-1.32)
0.74
 2151,9853,242,39514,35685,3861.05
(1.02-1.09)
78212,7481.28
(1.19-1.38)
 ≥ 3111,1842,307,93514,072

74,843

1.11
(1.07-1.15)
69711,1871.31
(1.21-1.41)
  1. HR, hazard ratio; CI, confidence interval

  2. *Adjusted for attained age and sex of the child, interaction between sex of the child and cancer of the parents, number of siblings, gestational age, mode of delivery and birth weight of the child, paternal age at child's birth, maternal age at child's birth, maternal smoking during early pregnancy, and the highest educational level of the parents.

  3. Adjusted for attained age of the child, interaction between attained age of the child and cancer of the parents, sex, number of siblings, gestational age, mode of delivery and birth weight of the child, paternal age at child's birth, maternal age at child's birth, maternal smoking during early pregnancy, and the highest educational level of the parents.

  4. Adjusted for attained age, sex and number of siblings of the child, interaction between number of siblings of the child and cancer of the parents, gestational age, mode of delivery and birth weight of the child, paternal age at child's birth, maternal age at child's birth, maternal smoking during early pregnancy, and the highest educational level of the parents

  5. §p = 0.054

Among all hospital contacts, 96% were due to unintentional injuries (HR, 1.07 [95% CI 1.05–1.09]). Parental cancer also tended to be associated with a higher rate of intentional self-harm (HR, 1.09 [95% CI 0.95–1.25]) and undetermined or other injuries (HR, 1.11 [95% CI 0.98–1.26]), but not of assault (HR, 0.99 [95% CI 0.87–1.13]). The associations did not appear to further differ by nature, body region, or mechanism of injury, or by place of injury occurrence, either during the entire follow-up or during the first year after cancer diagnosis (Figure 1).

Hazard ratios for hospital contacts for injury among children with parental cancer compared to children without parental cancer, according to different characteristics of injury (Hazard ratios were adjusted for attained age, sex, number of siblings, gestational age, mode of delivery and birth weight of the child, paternal age at child's birth, maternal age at child's birth, maternal smoking during early pregnancy, and the highest educational level of the parents).
https://doi.org/10.7554/eLife.08500.006

Among all events of injury, outpatient visit and hospitalization accounted for 83.5% and 16.5% respectively. Although the positive association was only statistically significant for outpatient visit during the entire follow-up (outpatient visit HR, 1.08 [95% CI 1.06–1.10]; hospitalization HR, 1.03 [95% CI 0.99–1.08]), the association was statistically significant for both hospitalization (HR, 1.18 [95% CI 1.07–1.31]) and outpatient visit (HR, 1.29 [95% CI 1.24–1.35]) during the first year after cancer diagnosis.

Secondary analysis

With 7-day washout periods, the mean number of hospital contacts for injury was 1.8 during the study period. Among children with one previous hospital contact, parental cancer was associated with a 1.24-fold rate of having a second hospital contact (95% CI 1.20–1.28) (Table 4). Similar patterns were observed for children with 2–4 previous injuries (Table 4). When all injuries were studied, children with parental cancer had a 13% higher rate of repeated injuries (HR, 1.13 [95% CI 1.12–1.15]). Additional analyses with 14-day and 30-day washout periods showed similar results (14-day HR, 1.12 [95% CI 1.10–1.13]; 30-day HR, 1.10 [95% CI 1.08–1.12]).

Table 4

Hazard ratios for hospital contact for injury among children with parental cancer compared to children without parental cancer, according to the number of previous hospital contact for injury of the child

https://doi.org/10.7554/eLife.08500.007
CharacteristicsNo. of Children With a Hospital Contact for InjuryPerson-yearsHR (95%CI) *
No contact
 No parental cancer548,48811,879,0751
 Parental cancer15,377298,3021.07 (1.05-1.09)
One contact
 No parental cancer228,5601,542,9261
 Parental cancer5,98131,3081.24 (1.20-1.28)
Two contacts
 No parental cancer107,833454,0491
 Parental cancer2,7969,1301.26 (1.20-1.32)
Three contacts
 No parental cancer53,784176,2371
 Parental cancer1,3773,6911.21 (1.13-1.31)
Four contacts
 No parental cancer28,12674,4991
 Parental cancer7221,6001.19 (1.07-1.33)
  1. HR, hazard ratio; CI, confidence interval

  2. *Adjusted for attained age, sex, number of siblings, gestational age, mode of delivery and birth weight of the child, paternal age at child's birth, maternal age at child's birth, maternal smoking during early pregnancy, and the highest educational level of the parents

Discussion

In this nationwide register-based study, we found that children having a parent with cancer had a higher rate of hospital contact for injury compared with other children. The rate increment was noted for children of all ages as well as for different kinds of injuries or places of injury occurrence, but was most pronounced immediately after the parent’s cancer diagnosis and among children with previous injuries. Comorbid psychiatric diagnoses after the cancer diagnosis rendered further higher rate increment of childhood injury.

Although it has been suggested that adolescents are most prone to psychosocial problems at the time of stressful life experience, younger children are in greater need of supervision and parenting (Phillips, 2014; Macpherson and Emeleus, 2007a; MacPherson and Emeleus, 2007b). The positive association between parental cancer and hospital contact for injury among children at all ages in the present study may therefore be jointly attributable to both the psychological distress among the children and the potential lack of parental supervision needed for injury prevention (Morrongiello et al., 2006a; Davis Kirsch et al., 2003; Faulkner and Davey, 2002; Asbridge et al., 2014; Bylund Grenklo et al., 2013). It has been debated whether boys and girls are differently affected by parental cancer (Krattenmacher et al., 2012; Visser et al., 2005). Our findings show that overall boys had a more pronounced rate increment than girls for injury. At the same age, boys are on average less mature in terms of social-emotional functioning compared to girls (Visser et al., 2005). However, worth noting is that despite the overall difference, boys and girls had similarly increased rates of injury, during the first year after parental cancer diagnosis.

Parental cancer was associated with a higher rate of injury, regardless of nature, mechanism, body region of the injury, or place of injury occurrence. Although a positive association was mainly noted for unintentional injuries, the lack of statistical significance for intentional injuries might be due to the relatively small number of intentional injuries observed. Interestingly and reassuringly, we found no increased rate of assault-related injuries after parental cancer. The fact that the higher rate of injury was noted not only at home, but also in transportation areas, in sports areas, etc., suggests that efforts in preventing injuries in children living with a parent with cancer should include a larger circle of support.

Children’s adjustment appears to vary at different stages of their parent’s cancer disease (Nelson and While, 2002; Huizinga et al., 2010). Our results showed clearly that children had the highest injury rate increase during the first year after the parent’s cancer diagnosis. This finding corroborates earlier findings in indicating that a cancer diagnosis poses severe psychological distress immediately after the diagnosis, both among the cancer patients and among their children (Fang et al., 2012; Lu et al., 2013; Fall et al., 2009; Huizinga et al., 2010). Previous studies have demonstrated that the well-being of children living with a parent with cancer is largely dependent on the adjustment status of their parents to the cancer (Krattenmacher et al., 2012; Nelson and While, 2002; Thastum et al., 2009; Huizinga et al., 2011). This was supported in our findings that children whose parent was also diagnosed with a psychiatric disorder after cancer diagnosis, appeared to have more pronounced rate increase of injury. Furthermore, among children with higher baseline risk of injury (i.e., children with previous hospital contact for injury), parental cancer was associated with an even more elevated risk for future injuries. These results highlight both a high-risk time window and high-risk groups for potential future interventions. Although the number of children living with a parent of cancer will undoubtedly increase due to the increasing cancer incidence and improving cancer survival, the postponement of childbearing, etc., the proportion of such children is however still small, making dedicated intervention both feasible and viable.

Conflicting results have been reported regarding whether maternal cancer has greater adverse impact on children than paternal cancer (Visser et al., 2005; Compas et al., 1994). In line with our previous finding on child mortality after parental cancer, the present study indicated no difference between maternal and paternal cancer in relation to the consequent risk of childhood injury (Chen et al., 2015). In contrast to previous findings, we found no difference in child injury risk by the severity of parental cancer (Krattenmacher et al., 2012). One potential explanation may be the fact that the severity of cancer does not always positively correlate with the adjustment status of the cancer patients. For example, in a recent large-scale study, it was reported that patients of cancers of relatively better survival (e.g., breast cancer) had the highest prevalence of mental disorders, whereas patients of cancers with much severe prognosis (e.g., pancreatic cancer) had the lowest prevalence (Mehnert et al., 2014).

This study is the first to use a population-based sample to examine the impact of parental cancer on the risk of childhood injury. The major strengths of our study include the nationwide cohort design, using the effective record linkage across the high quality Swedish population and health registers and the prospectively and independently collected information on exposure and outcome. These strengths enhance clearly the validity and generalizability of our findings. Some limitations of our study also deserve consideration. For instance, we had no information on the cohabitation or employment status of the parents. A cancer diagnosis may have considerable impact on the marital relationship and the family's economic status (Wozniak and Izycki, 2014; de Boer et al., 2009) which may in turn trigger additional psychological distress of the parents, leading to suboptimal parenting (Tein et al., 2000; Sallinen et al., 2004). Divorce or separation may further contribute to child injuries due to the departure of one parent from the household or simply a joint custody of the child between the parents. Therefore potential modifying effect of residence status within the family, as well as the cohabitation and employment status of the parents on the studied association deserves further investigation. Residual confounding due to unmeasured or unknown confounders is possible, however, with presumably small impact. In the multivariable models, only adjustment for the age of the child and the parental ages at child’s birth had noticeable impact on the increase in injury risk, whereas adjustment for other covariates including birth characteristics of the child and educational level of the parents, which have previously been suggested to be associated with both injury risk among the children and cancer risk among the parents, had rather negligible impact (data not shown) (Innes and Byers, 2004; Sun et al., 2010; Davey Smith et al., 2007; Beiki et al., 2014; Hemminki and Li, 2003). The facts that the rate increment was mainly noted during the first three years after parental cancer diagnosis but not thereafter, and that the rate increment was independent of number of siblings or whether the cancer is smoking/alcohol-related or not, further alleviated concerns about residual confounding. Misclassification of injuries remains possible as only above 80% of outpatient visits were included in the Patient Register currently (National Board of Health and Welfare, 2009). However, such misclassification is largely administrative and arguably non-differential. Cancer parents with established contact with health care may be more likely to seek medical care for their children’s injury. Yet, the opposite can also be postulated that while coping with this major illness, parents are less likely to seek medical care for minor hassles of their children. Such misclassification, if it exists, should have little impact on inpatient care for injury – a proxy of more severe injury event – and could not explain the largely increased injury rate during the first three years after parental cancer diagnosis whereas not thereafter.

In summary, children with a parent of cancer had a greater rate of hospital contact for injury, especially during the first year after cancer diagnosis. The association was also more pronounced for parental cancer with comorbid psychiatric disorders after the cancer diagnosis and among children with previous injuries.

Materials and methods

Study participants

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We conducted a historical cohort study from 2001 to 2010, including all children born in Sweden during 1983–2002 (n = 2,071,380) based on the Swedish Multi-Generation Register. The Swedish Multi-Generation Register contains information on all residents in Sweden who were born from 1932 onward and alive in 1961, together with their parents (Statistics Sweden, 2011). To be included in the present study, a child must have both biological parents alive, free of cancer and identifiable from this register before the child’s birth (n = 2,027,863).

Parental cancer

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All parents of these children were linked to the Swedish Cancer Register, which contains almost 100% complete information on all newly diagnosed cancer cases in Sweden since 1958 (Barlow et al., 2009). Information on type of cancer and date of diagnosis was collected from this register. If both parents were diagnosed with a cancer, the first diagnosis was used.

Childhood injury

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A hospital contact for injury was identified as either a hospitalization or an outpatient visit with injury according to the Swedish Patient Register. This register was initiated in 1964/1965 and has national coverage for hospital discharge records since 1987 (Ludvigsson et al., 2011). Since 2001, it also collects information on hospital-based outpatient specialist visits with over 80% coverage of the entire country (Ludvigsson et al., 2011). Information collected includes dates of admission and discharge, primary as well as multiple secondary diagnoses, and additionally external causes of morbidity and mortality when applicable. All diagnoses and external causes are coded according to Swedish revisions of the International Classification of Diseases (ICD). Since we were primarily interested in non-medical injuries, injuries due to complications of medical and surgical care were excluded from the definition of childhood injury in the present study. Thus, to be defined as a hospital contact for injury, the record had to have a main discharge diagnosis of injury (ICD 10: S00-T98 except T80-T88, T98.3) and an external cause (ICD 10: V01-Y98 except Y40-Y84, Y88).

In the primary analysis, we used the first hospital contact for injury during follow-up as the outcome and the date of admission or outpatient visit as the date of injury occurrence. To further examine whether the impact of parental cancer diagnosis differed between any hospital contact for injury (i.e., first hospital contact) and repeated hospital contacts for injury, we analyzed children that had more than one hospital contact for injury during the study period. In this secondary analysis, all hospital visits within a 7-day time period (wash-out period) was counted as one contact (i.e., more likely referring to the same injury). In additional analyses, we also used 14-day and 30-day wash-out periods to assess the robustness of this definition.

Follow-up

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In the primary analysis, all children were followed from January 1, 2001 or date of birth, whichever came later. Children without parental cancer contributed person-time to the unexposed period, whereas children with parental cancer contributed person-time first to the unexposed period and after date of parental cancer diagnosis to the exposed period. Children who had a parent diagnosed with cancer before January 1, 2001 contributed all person–time to the exposed period. For both exposed and unexposed periods, the follow-up was censored on the date of first hospital contact for injury, emigration, death, 18th birthday, or December 31, 2010, whichever occurred first. As a result, 63,236 children who had died or emigrated or became 18 years old before/at the start of follow-up were excluded, leaving 1,964,627 children in the final analyses.

In the secondary analysis, we specifically followed children (both exposed and unexposed) who already had one hospital contact for injury, from the end of wash-out period to the following injuries. For example, to examine the association of parental cancer diagnosis and a future injury among children that had already one hospital visit for injury, we followed all children with a first hospital contact for injury to the second one. Similar follow-ups were conducted when examining the risk of a third, fourth, etc. hospital contact for injury.

Covariates

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Various characteristics in children and parents have been linked to both risks of child injury and parental cancer and therefore might either confound or modify the studied association (Boutsikou and Malamitsi-Puchner, 2011; Morrongiello et al., 2007; Bradbury et al., 1999; Peden et al., 2008; Innes and Byers, 2004; Hjern, 2012; Weitzman et al., 1992; Weitoft et al., 2003; Sun et al., 2010; Hemminki and Li, 2003). To address potential confounding and effect modification, we collected information on sex, gestational age, mode of delivery and birth weight of the child, maternal smoking during early pregnancy and maternal age at child’s birth from the Swedish Medical Birth Register, as well as number of full and half siblings of the child and paternal age at child’s birth through the Multi-Generation Register. The Medical Birth Register was established in 1973 and has covered over 99% of all births in Sweden since 1983 (Centre for Epidemiology, National Board of Health and Welfare, 2003). We further identified the highest educational level of the parents from the Swedish Register of Education (Statistics Sweden. 2004).

Data availability

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The summary of data included in different registers used in the present study can be found on the homepages of the Swedish National Board of Health and Welfare (http://www.socialstyrelsen.se/register) as well as the Statistics Sweden (http://www.scb.se/sv_/Vara-tjanster/bestalla-mikrodata/Vilka-mikrodata-finns/).

The authors confirm that, for approved reasons, some access restrictions apply to the data underlying the findings. The data used in this study are owned by the Swedish National Board of Health and Welfare and Statistics Sweden. According to Swedish law, the authors are not able to make the dataset publicly available.

Any researchers (including international researchers) interested in obtaining the data can do so by the following steps: 1) apply for ethical approval from their local ethical review boards; 2) contact the Swedish National Board of Health and Welfare and/or Statistics Sweden with the ethical approval and make a formal application of use of register data. Contact emails for request of register data: Swedish National Board of Health and Welfare: registerservice@socialstyrelsen.se, Statistics Sweden: Mikrodata.individ@scb.se.

Please visit http://www.socialstyrelsen.se/register/bestalladatastatistik/bestallaindividuppgifterforforskningsandamal (the Swedish National Board of Health and Welfare) and http://www.scb.se/sv_/Vara-tjanster/bestalla-mikrodata/ (the Statistics Sweden) for detailed information about how to apply for access to register data for research purposes.

Statistical analysis

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Pearson’s χ2 test was used to compare the distributions of different child’s and parental characteristics between the exposed and unexposed children.

Primary analysis

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Cox proportional hazards regression was used to compare the rate of first hospital contact for injury between children with and without parental cancer. HR with 95% CI was estimated after adjustment for the covariates described above. To account for the correlation among children of the same parents, we used “clustered” (sandwich) standard errors in all models. Time since birth was used as the underlying time scale in the Cox models; no statistically significant violation of the proportional hazards assumption was detected from a test of the Schoenfeld residuals. Parental cancer diagnosis was treated as a time-varying exposure.

To examine the specific impact of cancer diagnosis, independent of the later course of the disease, we calculated the HRs of first hospital contact for injury during the first year, >1 and ≤3 years, and >3 years after parental cancer diagnosis separately. Children with a parental cancer diagnosed before start of follow-up might not contribute to all three categories, depending on when the parental cancer was diagnosed and when follow-up was censored. Since we used time since birth as the underlying time scale, different HRs observed from these analyses did not conflict with the proportional hazards assumption tested. We sub-grouped parental cancer to explore whether maternal and paternal cancer had a different impact on child injury. To assess the impact of lifestyle factors as potential confounders for the studied association, we sub-grouped parental cancer as tobacco-related and other cancers, or alcohol-related and other cancers (National Board of Health and Welfare, 2013; World Health Organization). To explore the potential modifying effect of cancer severity, we further categorized parental cancer as cancer with high, medium or low expected 5-year survival. The expected 5-year survival was indexed as the predicted 5-year relative survival rates of different cancer types based on the entire Cancer Register (Talback et al., 2004). We further ascertained from the Patient Register hospital contacts for selected psychiatric comorbidity that were newly diagnosed after the cancer diagnosis among the parents. The psychiatric diagnoses considered were depression, anxiety disorders, stress reaction and adjustment disorder (detailed diagnoses and corresponding ICD codes are listed in the Table 5). We performed Wald tests to compare the HRs for different subgroups.

Table 5

Swedish revisions of the international classification of diseases (ICD) for psychiatric comorbidity of the cancer parents.

https://doi.org/10.7554/eLife.08500.008
ICD 8 (1969-1986)ICD 9 (1987-1996)ICD-10 (1997-presesnt)
Depression296.2, 298.0, 300.4296B, 300E, 311F32-F39
Anxiety disorders300 except 300.3, 300.4300 except 300D, 300EF40, F41, F44, F45, F48
Stress reaction and adjustment disorder307308, 309F43

To assess whether the impact of parental cancer on child injury differed by sex, age or number of siblings of the child, we used formal tests of interaction of parental cancer with sex, age at follow-up (<3, 3–5, 6–11, 12–15 or ≥ 15 years), or number of full and half siblings (0, 1, 2, ≥3) of the child.

To examine whether the studied association differed for different types of injury, we further conducted separate analyses by manner or intent, nature, body region and mechanism of injury, as well as by place of injury occurrence. To assess whether the association varied by different severity of injury, we also examined separately the risk of hospitalization and outpatient visit for injury.

Secondary analysis

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Ordinary Cox proportional hazard regression was used to assess the association between parental cancer diagnosis and a new injury among children with at least one previous hospital contact for injury during follow-up. A conditional Cox model (PWP-TT model) was used to assess the overall association between parental cancer diagnosis and repeated injuries in children (Amorim and Cai, 2015).

For all analyses, statistical significance was assessed using 2-tailed 0.05-level tests. Data preparation was performed using SAS version 9.4. Statistical analyses were performed using Stata version 12.1.

The study was approved by the Central Ethical Review Board in Stockholm, Sweden. All individuals' information was anonymized and de-identified prior to analysis.

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Decision letter

  1. Eduardo Franco
    Reviewing Editor; McGill University, Canada

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.

Thank you for submitting your work entitled "Childhood injury after a parental cancer diagnosis" for peer review at eLife. We are pleased to inform you that your submission has been favorably evaluated by Prabhat Jha (Senior editor) and two reviewers, one of whom (Eduardo Franco) is a member of our Board of Reviewing Editors.

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

The effects of serious parental illness on children's injuries is an interesting topic with potential health policy implications. The study is based on good quality data, and the research questions are clearly motivated.

1) Intuitively, you could have conducted a simpler study, e.g., a case-control study resorting to data linkage to the Swedish Cancer Registry. Instead, you conducted a much stronger cohort study using the opportunities for effective record linkage with administrative and healthcare databases in Sweden. It seems to be the first study of this kind. Please underscore this fact briefly in the Discussion, i.e., the enhanced validity that presumably came from this more robust design.

2) Your contention that the state of cohabitation (unmeasured possible confounder) could be a problem is well reasoned. It is conceivable that there were more divorces around the time of cancer diagnosis. Cancer is devastating to a couple's wellbeing and may lead to a situation that triggers a separation, which in turn may lead to neglect by the caregiving parent in exerting proper surveillance to the child. It is also possible that the divorce will have been the triggering stressful event preceding a cancer diagnosis (a typical anecdote that oncologists like to recall). Either way, the ensuing separation may further contribute to injuries due to the departure of one of the parents from the household. Another scenario favoring injuries is also consistent with the above conditions. Separated or divorced parents may have joint custody of the child, which entails travel between two households and exposure to unfamiliar environments for the child, which in turn could lead to injuries. You should consider the extent of these possible conditions with respect to the social and behavioral mechanisms behind their findings. Please expand the Discussion to elaborate on these possible mechanisms.

3) Although the number of children that are exposed to parental cancer are undoubtedly increasing (one reason being postponement of childbearing), the proportion is still very small. Please acknowledge this fact in the Discussion. Consider also adding a percentage to the number of exposed (Results, first paragraph).

4) The linking of parents and children is not entirely clear. You seem to refer to parents as biological parents of the children. You did not consider whether the children actually reside with their parents (and siblings). This is relevant with regard to the exposure to an ill parent and should be better acknowledged.

5) Expand on the reasons for your choice of covariates in the multivariate models, especially in the Discussion where you noted that most covariates had negligible impact (sixth paragraph). For example, although statistically significant, the differences with regard to birth weight, gestational age and mode of delivery were very small. If the tendency of adjustments was to render the associations stronger, it would be useful to know which covariates made the difference, which would provide insights as to the mechanisms leading to injuries.

6) You state that the records for hospital care had to have a main discharge diagnosis of injury (ICD 10: S00-T98) and an external cause (ICD 10: V01-Y98). Does this mean that hospital care episodes due to complications of medical and surgical care were included? Considering the main motivation for this study, these episodes should have been excluded as the focus lies on the children's proneness for injuries.

7) The follow-up starts on 1.1.2001, however, you stated that children whose parent had an earlier diagnosis contributed all follow-up time to the exposed period (subsection “Follow-up”). How were these children treated in the analyses that used time since cancer diagnosis?

8) Did the risks vary at all by injury severity (subsection “Primary analysis”)? This is an interesting point as hospitalizations are presumably less affected by the parent's propensity to seek care for their children.

9) It is surprising that the risks did not differ by child's age as the typical injuries and their context do differ a lot between toddlers and teenagers. Would a different choice of age groupings have made a difference? For instance, the 12-18 group is quite heterogeneous in terms of dependence from parents. This is also the age group that contributes the vast majority of all events in the follow-up. Splitting this group could have been informative.

10) Make it clear in the Abstract that all the reported hazard ratios are adjusted for important confounders and covariates.

https://doi.org/10.7554/eLife.08500.013

Author response

1) Intuitively, you could have conducted a simpler study, e.g., a case-control study resorting to data linkage to the Swedish Cancer Registry. Instead, you conducted a much stronger cohort study using the opportunities for effective record linkage with administrative and healthcare databases in Sweden. It seems to be the first study of this kind. Please underscore this fact briefly in the Discussion, i.e., the enhanced validity that presumably came from this more robust design.

Thank you for the suggestion. We have now stressed this strength in the Discussion section (sixth paragraph).

2) Your contention that the state of cohabitation (unmeasured possible confounder) could be a problem is well reasoned. It is conceivable that there were more divorces around the time of cancer diagnosis. Cancer is devastating to a couple's wellbeing and may lead to a situation that triggers a separation, which in turn may lead to neglect by the caregiving parent in exerting proper surveillance to the child. It is also possible that the divorce will have been the triggering stressful event preceding a cancer diagnosis (a typical anecdote that oncologists like to recall). Either way, the ensuing separation may further contribute to injuries due to the departure of one of the parents from the household. Another scenario favoring injuries is also consistent with the above conditions. Separated or divorced parents may have joint custody of the child, which entails travel between two households and exposure to unfamiliar environments for the child, which in turn could lead to injuries. You should consider the extent of these possible conditions with respect to the social and behavioral mechanisms behind their findings. Please expand the Discussion to elaborate on these possible mechanisms.

We agree and have now added comments in this regard in the Discussion section (sixth paragraph).

3) Although the number of children that are exposed to parental cancer are undoubtedly increasing (one reason being postponement of childbearing), the proportion is still very small. Please acknowledge this fact in the Discussion. Consider also adding a percentage to the number of exposed (Results, first paragraph).

We agree and have now added comments in this regard in the Discussion section (fourth paragraph). We also as suggested added a percentage to the number of exposed children in the Results section (first paragraph).

4) The linking of parents and children is not entirely clear. You seem to refer to parents as biological parents of the children. You did not consider whether the children actually reside with their parents (and siblings). This is relevant with regard to the exposure to an ill parent and should be better acknowledged.

We indeed only studied cancer diagnosis among biological parents of the children. We have now added more information regarding the Swedish Multi-Generation Register and clarified that we only included children with both biological parents identifiable from this register in the present study (subsection “Study participants”).

We also agree that it is important to acknowledge that in the present study we were unable to identify whether or not the child was actually residing with the parents at the time of parental cancer diagnosis. We have added comments in this regard in the Discussion section (sixth paragraph).

5) Expand on the reasons for your choice of covariates in the multivariate models, especially in the Discussion where you noted that most covariates had negligible impact (sixth paragraph). For example, although statistically significant, the differences with regard to birth weight, gestational age and mode of delivery were very small. If the tendency of adjustments was to render the associations stronger, it would be useful to know which covariates made the difference, which would provide insights as to the mechanisms leading to injuries.

In the multivariable models, only adjustment for child age at follow-up and parental ages at cancer diagnosis had noticeable impacts on the hazard ratios (HRs) for childhood injuries, whereas adjustment for other covariates had negligible effects. The crude incidence rate ratios for childhood injury any time after parental cancer diagnosis and during the first year after parental cancer diagnosis were 1.12 (95% CI: 1.10-1.13) and 1.30 (95% CI: 1.25-1.35) respectively. For the corresponding HRs only adjusted for child age, adjusted for both child age and parental ages at child birth and the fully adjusted HRs, please refer to the below table.

HR (95% CI) adjusted for child age

HR (95% CI) adjusted for child age and parental ages at child birth

Fully adjusted HR (95% CI)

Any time after parental cancer diagnosis

1.04 (1.02-1.06)

1.08 (1.06-1.09)

1.07 (1.05-1.09)

First year after parental cancer diagnosis

1.24 (1.19-1.28)

1.28 (1.23-1.33)

1.27 (1.22-1.33)

We have now further clarified the choice of covariates in the Discussion section (sixth paragraph).

6) You state that the records for hospital care had to have a main discharge diagnosis of injury (ICD 10: S00-T98) and an external cause (ICD 10: V01-Y98). Does this mean that hospital care episodes due to complications of medical and surgical care were included? Considering the main motivation for this study, these episodes should have been excluded as the focus lies on the children's proneness for injuries.

Thank you for this very good comment. We have now excluded injuries related to complications of surgical and medical care, i.e. medical injuries defined as injuries with a main discharge diagnosis of ICD 10 codes T80-T88 or T98.3 or with an external cause of ICD 10 codes Y40-Y84 or Y88, from the outcome definition of the analyses. As we expected, among this healthy study population, such injuries comprised a rather small proportion of all injuries identified (among unexposed children, 1.2%, and among exposed children, 1.2%). We have now modified the definition of childhood injury in the Materials and methods section (subsection “Childhood injury”), and updated all results in the Abstract section, Results section (subsections “Primary analysis” and “Secondary analysis”), Figure 1, Table 2, Table 3 and Table 4.

7) The follow-up starts on 1.1.2001, however, you stated that children whose parent had an earlier diagnosis contributed all follow-up time to the exposed period (subsection “Follow-up”). How were these children treated in the analyses that used time since cancer diagnosis?

In the analyses taken into account the entire follow-up, children whose parent had a cancer diagnosis before January 1, 2001 contributed all person–time (i.e. person-time experienced from January 1, 2001 to the end of follow-up) to the exposed period. In the analyses where we examined the impact of time since parental cancer diagnosis on childhood injury, we split children’s exposed period into 3 periods, i.e. 1 year or less, >1 and ≤3 years, more than 3 years, after parental cancer diagnosis. For example, if the parent had a cancer diagnosis on January 1, 1999 and the child was followed from January 1, 2001 until December 31, 2010, this child would contribute 1 person-year (January 1, 2001–December 31, 2001) to the “>1 and ≤3 years” category and 9 person-years (January 1, 2002 – December 31, 2010) to the “more than 3 years” category in the analyses. We have now added additional comments in this regard in the Materials and methods section (subsection “Primary analysis”).

8) Did the risks vary at all by injury severity (subsection “Primary analysis”)? This is an interesting point as hospitalizations are presumably less affected by the parent's propensity to seek care for their children.

We used the type of hospital contact as a proxy of injury severity, i.e., hospitalization/inpatient visit=more severe, outpatient visit=less severe, and calculated HRs for hospitalization and outpatient visit for injury separately. To compare the HRs of hospitalization and outpatient visit for injury, no direct method was available to our best knowledge. We therefore used a general method for comparing two HRs, exp(β1) and exp(β2), as previously reported (Altman and Bland, 2003). In brief, we first calculated z score= (β1– β2) / [SE(β1)2 +SE(β2)2], then the calculated z score was compared to the standard normal distribution, leading to a P value. If P<0.05, we report that the compared two HRs are statistically significantly different. In the present study, the HR of hospitalization for injury during the entire follow-up was 1.03 (95% CI: 0.99-1.08) and the HR of outpatient visit for injury was 1.08 (95% CI: 1.06-1.10) (section "Results"), leading to a P value of 0.10. The HR of hospitalization for injury during the first year after parental cancer diagnosis was 1.18 (95% CI: 1.07-1.31) and the HR of outpatient visit for injury during the first year was 1.29 (95% CI: 1.24-1.35) (section "Results"), leading to a P value of 0.20. Although the overall association tended to be stronger for outpatient visit for injury which would support the impact of parent’s healthcare-seeking propensity, the difference was not statistically significant for either any time or the first year after cancer diagnosis. We therefore concluded that we could not draw a conclusion regarding whether or not the HRs for hospitalization and outpatient visit for injury were statistically significantly different from each other. A presumption of this method is that the two compared HRs are estimated from different (independent) samples. In the case of overlapping samples, as in the present study, this formula should be modified by adding a covariance term for β1 and β2 in the denominator. Although we generally presume that the covariance term is relatively small and has little influence on the estimated P values, it is unfortunately rather hard to precisely estimate this covariance term according to our knowledge.

The fact that both hospitalizations and outpatient visits for childhood injuries clearly increased during the first year after parental cancer diagnosis (18% for hospitalization and 29% for outpatient visit) and that the increased risk was limited to the first three years after diagnosis whereas not thereafter, further argued against a pure explanation by differential healthcare-seeking behavior between the affected parents and other parents. We have modified the discussion about this in the Discussion section (sixth paragraph).

9) It is surprising that the risks did not differ by child's age as the typical injuries and their context do differ a lot between toddlers and teenagers. Would a different choice of age groupings have made a difference? For instance, the 12-18 group is quite heterogeneous in terms of dependence from parents. This is also the age group that contributes the vast majority of all events in the follow-up. Splitting this group could have been informative.

Thank you for the suggestion. We have further split the 12-18 years group into two groups, i.e. 12-15 years group and 15-18 years group and reported the corresponding HRs with 95% CIs. It seems that there was no evidence that younger and older adolescents had different risks of hospital contacts for injury when a parent had a cancer diagnosis. We have added information in this regard in the Materials and methods section (subsection “Primary analysis”) and presented new results in Table 3.

10) Make it clear in the Abstract that all the reported hazard ratios are adjusted for important confounders and covariates.

We have revised the Abstract accordingly.

https://doi.org/10.7554/eLife.08500.014

Article and author information

Author details

  1. Ruoqing Chen

    Department of Medical Epidemiology and Biostatistics, Karolinska Institutet, Stockholm, Sweden
    Contribution
    RC, Obtained funding, Conception and design, Analysis and interpretation of data, Drafting or revising the article
    Contributed equally with
    Amanda Regodón Wallin
    For correspondence
    ruoqing.chen@ki.se
    Competing interests
    The authors declare that no competing interests exist.
    ORCID icon "This ORCID iD identifies the author of this article:" 0000-0003-4911-3543
  2. Amanda Regodón Wallin

    Department of Medical Epidemiology and Biostatistics, Karolinska Institutet, Stockholm, Sweden
    Contribution
    ARW, Conception and design, Analysis and interpretation of data, Drafting or revising the article
    Contributed equally with
    Ruoqing Chen
    Competing interests
    The authors declare that no competing interests exist.
  3. Arvid Sjölander

    Department of Medical Epidemiology and Biostatistics, Karolinska Institutet, Stockholm, Sweden
    Contribution
    AS, Conception and design, Analysis and interpretation of data, Drafting or revising the article
    Competing interests
    The authors declare that no competing interests exist.
  4. Unnur Valdimarsdóttir

    1. Department of Medical Epidemiology and Biostatistics, Karolinska Institutet, Stockholm, Sweden
    2. Center of Public Health Sciences, Faculty of Medicine, University of Iceland, Reykjavík, Iceland
    Contribution
    UV, Conception and design, Drafting or revising the article
    Competing interests
    The authors declare that no competing interests exist.
  5. Weimin Ye

    Department of Medical Epidemiology and Biostatistics, Karolinska Institutet, Stockholm, Sweden
    Contribution
    WY, Obtained funding, Conception and design, Acquisition of data, Drafting or revising the article
    Competing interests
    The authors declare that no competing interests exist.
  6. Henning Tiemeier

    1. Department of Epidemiology, Erasmus MC University Medical Center, Rotterdam, The Netherlands
    2. Department of Child and Adolescent Psychiatry, Erasmus MC University Medical Center, Rotterdam, The Netherlands
    Contribution
    HT, Conception and design, Drafting or revising the article
    Competing interests
    The authors declare that no competing interests exist.
    ORCID icon "This ORCID iD identifies the author of this article:" 0000-0002-4395-1397
  7. Katja Fall

    Clinical Epidemiology and Biostatistics, Faculty of Medicine and Health, Örebro University, Örebro, Sweden
    Contribution
    KF, Conception and design, Drafting or revising the article
    Competing interests
    The authors declare that no competing interests exist.
  8. Catarina Almqvist

    1. Department of Medical Epidemiology and Biostatistics, Karolinska Institutet, Stockholm, Sweden
    2. Lung and Allergy Unit, Astrid Lindgren Children’s Hospital, Karolinska University Hospital, Stockholm, Sweden
    Contribution
    CA, Obtained funding, Conception and design, Acquisition of data, Drafting or revising the article
    Competing interests
    The authors declare that no competing interests exist.
  9. Kamila Czene

    Department of Medical Epidemiology and Biostatistics, Karolinska Institutet, Stockholm, Sweden
    Contribution
    KC, Conception and design, Acquisition of data, Drafting or revising the article
    Competing interests
    The authors declare that no competing interests exist.
  10. Fang Fang

    Department of Medical Epidemiology and Biostatistics, Karolinska Institutet, Stockholm, Sweden
    Contribution
    FF, Obtained funding, Conception and design, Analysis and interpretation of data, Drafting or revising the article
    Competing interests
    The authors declare that no competing interests exist.

Funding

Forskningsrådet för Hälsa, Arbetsliv och Välfärd (2012-0498)

  • Fang Fang

China Scholarship Council (201206100002)

  • Ruoqing Chen

Svenska Sällskapet för Medicinsk Forskning (Researcher position)

  • Fang Fang

Karolinska Institutet (Funding for Strategic Young Scholar Grants in Epidemiology)

  • Catarina Almqvist
  • Fang Fang

Karolinska Institutet (Assistant professor position)

  • Fang Fang

Vetenskapsrådet (SIMSAM 340-2013-5867)

  • Weimin Ye
  • Catarina Almqvist

Vetenskapsrådet (SIMSAM 80748301)

  • Weimin Ye
  • Catarina Almqvist

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

Acknowledgements

The authors wish to thank Dr Eva Norén Selinus for helpful conversations regarding diagnosis of psychiatric disorders in patients with cancer.

Ethics

Human subjects: The study was approved by the Central Ethical Review Board (Centrala etikprövningsnämnden) in Stockholm, Sweden (Dnr Ö 12-2013). In accordance with their decision, we did not obtain informed consent from participants involved in the study. All individuals' information was anonymized and de-identified prior to analysis.

Reviewing Editor

  1. Eduardo Franco, McGill University, Canada

Publication history

  1. Received: May 4, 2015
  2. Accepted: October 27, 2015
  3. Accepted Manuscript published: October 31, 2015 (version 1)
  4. Version of Record published: February 9, 2016 (version 2)

Copyright

© 2015, Chen 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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    Flavia Camponovo et al.
    Research Article Updated

    Tanzanian adult male volunteers were immunized by direct venous inoculation with radiation-attenuated, aseptic, purified, cryopreserved Plasmodium falciparum (Pf) sporozoites (PfSPZ Vaccine) and protective efficacy assessed by homologous controlled human malaria infection (CHMI). Serum immunoglobulin G (IgG) responses were analyzed longitudinally using a Pf protein microarray covering 91% of the proteome, providing first insights into naturally acquired and PfSPZ Vaccine-induced whole parasite antibody profiles in malaria pre-exposed Africans. Immunoreactivity was identified against 2239 functionally diverse Pf proteins, showing a wide breadth of humoral response. Antibody-based immune ‘fingerprints’ in these individuals indicated a strong person-specific immune response at baseline, with little changes in the overall humoral immunoreactivity pattern measured after immunization. The moderate increase in immunogenicity following immunization and the extensive and variable breadth of humoral immune response observed in the volunteers at baseline suggest that pre-exposure reduces vaccine-induced antigen reactivity in unanticipated ways.

    1. Epidemiology and Global Health
    Lauren C Tindale et al.
    Research Article Updated

    We collated contact tracing data from COVID-19 clusters in Singapore and Tianjin, China and estimated the extent of pre-symptomatic transmission by estimating incubation periods and serial intervals. The mean incubation periods accounting for intermediate cases were 4.91 days (95%CI 4.35, 5.69) and 7.54 (95%CI 6.76, 8.56) days for Singapore and Tianjin, respectively. The mean serial interval was 4.17 (95%CI 2.44, 5.89) and 4.31 (95%CI 2.91, 5.72) days (Singapore, Tianjin). The serial intervals are shorter than incubation periods, suggesting that pre-symptomatic transmission may occur in a large proportion of transmission events (0.4–0.5 in Singapore and 0.6–0.8 in Tianjin, in our analysis with intermediate cases, and more without intermediates). Given the evidence for pre-symptomatic transmission, it is vital that even individuals who appear healthy abide by public health measures to control COVID-19.