Abstract
OBJECTIVE:
To determine whether neonatal phototherapy is associated with cancer in the first year after birth.
METHODS:
We analyzed a data set from the California Office of Statewide Health Planning and Development that was created by linking birth certificates, death certificates, and hospital discharge abstracts up to age 1 year. Subjects were 5 144 849 infants born in California hospitals at ≥35 weeks’ gestation from 1998 to 2007. We used International Classification of Diseases, Ninth Revision codes to identify phototherapy at <15 days and discharge diagnoses of cancer at 61 to 365 days. We adjusted for potential confounding variables by using traditional and propensity-adjusted logistic regression models.
RESULTS:
Cancer was diagnosed in 58/178 017 infants with diagnosis codes for phototherapy and 1042/4 966 832 infants without such codes (32.6/100 000 vs 21.0/100 000; relative risk 1.6; 95% confidence interval [CI], 1.2–2.0, P = .002). In propensity-adjusted analyses, associations were seen between phototherapy and overall cancer (adjusted odds ratio [aOR] 1.4; 95% CI, 1.1–1.9), myeloid leukemia (aOR 2.6; 95% CI, 1.3–5.0), and kidney cancer (aOR 2.5; 95% CI, 1.2–5.1). The marginal propensity-adjusted absolute risk increase for cancer after phototherapy in the total population was 9.4/100 000 (number needed to harm of 10 638). Because of the higher baseline risk of cancer in infants with Down syndrome, the number needed to harm was 1285.
CONCLUSIONS:
Phototherapy may slightly increase the risk of cancer in infancy, although the absolute risk increase is small. This risk should be considered when making phototherapy treatment decisions, especially for infants with bilirubin levels below current treatment guidelines.
What’s Known on This Subject:
Phototherapy is commonly used to treat neonatal jaundice. It is generally thought to be safe. However, a small number of studies have suggested an association between neonatal phototherapy and cancer.
What This Study Adds:
In this study of ∼5 million California births, phototherapy was associated with a small but statistically significant increased risk of some infantile cancers, specifically myeloid leukemia and kidney cancer, with ∼1 additional cancer case per 10 638 treated infants.
Phototherapy is commonly used to treat neonatal jaundice. The use of phototherapy has been increasing1,2; a recent study of term and near-term infants enrolled in a single health plan in California showed that up to 15.9% received this treatment.3 Many clinicians initiate phototherapy at levels lower than those recommended by the American Academy of Pediatrics.2,4 The rise in phototherapy use may be a result of increased identification of neonates with hyperbilirubinemia through universal screening protocols,2 fear of kernicterus,5 and general belief that phototherapy is safe.4
Although phototherapy is thought to pose minimal risk to infants with unconjugated hyperbilirubinemia,4,6 2 large epidemiologic studies have found phototherapy to be associated with subsequent cancer (overall leukemia and acute myelogenous leukemia).7,8 These associations have not been consistent; some studies have been reassuring,9,10 but none have been large enough to be conclusive. The potential for phototherapy to increase the incidence of cancer is supported by data from in vitro and in vivo studies.11–22
It is important to determine whether phototherapy contributes to early childhood cancer to determine the optimal management of infants with neonatal hyperbilirubinemia. In this study, we used population-based data to investigate the association between phototherapy and cancer in the first year after birth.
Methods
Study Design and Subjects
This study is an analysis of a historical cohort of infants born in California between January 1, 1998 and December 31, 2007. We used the linked Vital Statistics/Patient Discharge Dataset (VS/PDD) for our analyses. The VS/PDD was created for the California Office of Statewide Health Planning and Development by using probabilistic linkage of data from birth certificates, death certificates, and hospital discharge abstracts for mothers and their newborn infants up to 1 year of age.23 The VS/PDD includes all California births except those of infants born at home or in military hospitals. For our analyses, we excluded infants born at <35 weeks’ gestation and infants who died within 60 days of birth.
This study was approved by the institutional review boards for the protection of human subjects at the California Health and Human Services Agency, California Department of Public Health, and the University of California, San Francisco.
Predictor Variables
The primary predictor variable was receipt of phototherapy during a hospitalization that began within the first 2 weeks after birth. We ascertained phototherapy use by abstracting International Classification of Diseases, Ninth Revision (ICD-9)24 procedure codes for phototherapy (99.82, 99.83).
Covariates abstracted from the birth certificates included gender, birth weight, gestational age, twin birth, birth by cesarean delivery, payer source, year of birth, and maternal and paternal demographic variables (race, Hispanic ethnicity, age, and education). Covariates abstracted from discharge diagnosis ICD-9 codes included jaundice (774.0–774.4, 774.5, 774.6, 782.4), Down syndrome (758.0), other chromosomal anomalies (758.1–758.9), and nonchromosomal congenital anomalies (740.0–743.64, 743.66–749.25, 750.1–757.32, 757.39–757.9, 759.0–759.9); only congenital anomalies diagnosed during the birth hospitalization were included.
Data were missing for several of our covariates, most commonly payer source (missing in 12.1%), paternal education (9.4%), gestational age (7.1%), paternal age (7.1%), and paternal race (6.9%). We imputed missing values by using single regression imputation because of the large size of the data set. To estimate the effect of the missing data on the precision of our results, we performed multiple imputation (5 imputations) on a subset that included all cancer cases and a 1% sample of the remainder of the cohort. We included subjects with missing gestational age if their imputed gestational age (based on multiple variables including birth weight, parental race, parental age, mode of delivery, and twin status) was ≥35 weeks.
Outcome Variables
The primary outcome was a hospital discharge diagnosis of cancer (ICD-9 codes 140.0–209.79, 230.0–234.9) in the first year after birth. We excluded cancers diagnosed at ≤60 days to reduce the likelihood of finding an association through reverse causation. In addition, it seemed less plausible that phototherapy would cause cancer in ≤60 days. We grouped cancer diagnosis ICD-9 codes by site: myeloid leukemia (205.0–205.92), lymphoid leukemia (204.0–204.92), brain or nervous system (191.0–192.9), kidney (189.0–189.9), liver (155.0–156.9), soft tissue (171.0–171.9), and other (remaining codes). If an infant had >1 type of cancer diagnosis, all were captured.
Statistical Analysis
We used Stata 13 (Stata Corp, College Station, TX) statistical software for all analyses. We performed bivariate analyses comparing infants with and without cancer diagnoses by using χ2 tests for categorical variables and t tests for continuous variables.
We computed risk ratios (RRs) and risk differences for phototherapy and each of the cancer outcomes by using the Agresti-Caffo method25 to compare risk differences. Because the exposure (phototherapy) is much more common than the outcome (various types of cancer), creating a model for exposure allowed control for many more potential confounding variables than traditional analyses that create a model for the outcome. We therefore created a propensity score for phototherapy with the following covariates: gender, birth weight, gestational age, large for gestational age, twin birth, birth by cesarean delivery, payer source, year of birth, maternal race, paternal race, maternal Hispanic ethnicity, maternal age, paternal age, maternal education, paternal education, Down syndrome, other chromosomal anomalies, and nonchromosomal congenital anomalies. The propensity score had an area under the receiver operating characteristic curve of 0.67. We then used logistic regression to evaluate the association between phototherapy and cancer, adjusted for the propensity to receive phototherapy. In the propensity-adjusted analyses, the predictor variables were only phototherapy and propensity to receive phototherapy (in categories). We calculated unadjusted and marginal propensity-adjusted risk differences and 95% confidence intervals (CIs) of the risk differences; we calculated marginal risk differences because they estimate effect in the entire population (ie, if all versus no infants were treated with phototherapy), whereas other risk differences depend on baseline risk (ie, the values of other covariates).26 We calculated numbers needed to harm by taking the reciprocal of the risk differences. To complement propensity-adjusted models, we also analyzed the data in traditional multivariable logistic models adjusted for gender, birth weight, gestational age, Down syndrome, maternal race, and year of birth. To ensure that imputation did not affect our results, we performed a complete case analysis.
Because Down syndrome is such a strong risk factor for cancer,27 we estimated the marginal propensity-adjusted risk difference and 95% CI for cancer after phototherapy in infants with and without Down syndrome separately. For this analysis, we created a propensity score without the Down syndrome variable.
Results
Our cohort consisted of 5 144 849 infants born at ≥35 weeks’ gestation who were alive >60 days after birth. A cancer diagnosis was present in 1100 of the infants, yielding an overall population incidence of 21.4/100 000. Table 1 provides a description of the cohort by cancer diagnosis status. The most common cancers were leukemia (23%), brain or nervous system cancer (16%), and eye or orbit cancer (9%).
TABLE 1.
Description of Cohort (n = 5 144 861), by Diagnosis of Cancer
| Cancer Diagnosis | P | ||
|---|---|---|---|
| Yes (n = 1100) | No (n = 5 143 749) | ||
| Male, % | 55.2 | 51 | .005 |
| Gestation in wk, mean (SD) | 39.5 (1.7) | 39.5 (1.6) | .65 |
| Birth wt in g, mean (SD) | 3411 (511) | 3385 (503) | .09 |
| Large for gestational age, %a | 10.9 | 9.9 | .28 |
| Twin, % | 2.4 | 2.2 | .74 |
| Down syndrome, % | 1.6 | 0.2 | <.001 |
| Other chromosomal anomalies, % | 0.6 | 0.1 | <.001 |
| Congenital anomalies, % | 7.7 | 4.1 | <.001 |
| Maternal race, % | .1 | ||
| White | 83 | 80.2 | |
| Black | 5 | 5.8 | |
| Asian | 10.2 | 11.4 | |
| Other | 1.9 | 2.6 | |
| Maternal Hispanic ethnicity, % | 50.6 | 47.5 | .04 |
| Paternal race, % | .36 | ||
| White | 82.4 | 80.2 | |
| Black | 5.8 | 6.7 | |
| Asian | 9.4 | 10.6 | |
| Other | 2.4 | 2.6 | |
| Maternal age in y, mean (SD) | 28.2 (6.4) | 27.9 (6.3) | .08 |
| Paternal age in y, mean (SD) | 30.9 (7) | 30.8 (7.1) | .58 |
| Maternal education, % | .78 | ||
| Did not complete high school | 31.4 | 29.8 | |
| Completed high school, no college | 26.3 | 27.4 | |
| Completed some college | 20.2 | 20 | |
| Completed 4 y of college | 12.6 | 13.1 | |
| Completed >4 y of college | 9.5 | 9.8 | |
| Paternal education, % | .7 | ||
| Did not complete high school | 30.4 | 28.5 | |
| Completed high school, no college | 29.2 | 29.2 | |
| Completed some college | 16.7 | 17.9 | |
| Completed 4 y of college | 12.8 | 13.2 | |
| Completed >4 y of college | 11 | 11.1 | |
| Cesarean delivery, % | 29.5 | 26.2 | .02 |
| Payer, % | .07 | ||
| Private insurance | 50.6 | 51.5 | |
| Medi-Cal | 46.7 | 44.6 | |
| Other | 2.7 | 4 | |
Large for gestational age is defined as >90th percentile (in this data set).
Jaundice diagnoses were recorded in 13.9% of infants and phototherapy codes in 3.5% of the cohort. Phototherapy use increased over time, from 2.9% in 1998 to 4.4% in 2007 (P < .001). Of subjects who received phototherapy, 85% had a jaundice diagnosis. Jaundice itself was associated with an increased risk of cancer, controlling for phototherapy (RR 1.3; 95% CI, 1.1–1.5).
In unadjusted analyses, subjects who received phototherapy were at higher risk of overall cancer (RR 1.6; 95% CI, 1.2–2.0), myeloid leukemia (RR 2.7; 95% CI, 1.4–5.2), kidney cancer (RR 2.7; 95% CI, 1.3–5.6), and other cancer (RR 1.7; 95% CI, 1.1–2.5), although the risk differences were small (Tables 2 and 3). The associations between phototherapy and overall cancer, myeloid leukemia, kidney cancer, and other cancer persisted in propensity-adjusted models (Table 3). Propensity-adjusted models, traditional multivariable-adjusted models, and models with and without multiple imputation gave similar results. In a complete case analysis (n = 3 717 503), we found similar propensity-adjusted associations between phototherapy and overall cancer (odds ratio [OR] 1.5; 95% CI, 1.1–2.1), myeloid leukemia (OR 2.2; 95% CI, 1.0–4.9), kidney cancer (OR 3.4; 95% CI, 1.6–7.3), and other cancer (OR 1.7; 95% CI, 1.1–2.7). After adjustment for jaundice alone, the association between phototherapy and myeloid leukemia was similar (OR 2.6; 95% CI, 1.2–5.6), the associations between phototherapy and overall cancer (OR 1.3; 95% CI, 1.0–1.8) and other cancer (OR 1.5; 95% CI, 1.0–2.4) were slightly diminished, and the association between phototherapy and kidney cancer (OR 1.6; 95% CI, 0.7–3.5) was diminished further. The marginal propensity-adjusted absolute risk increase for cancer after phototherapy in the total population was 9.4/100 000 (95% CI, 1.4/100 000–17.4/100 000), with a number needed to harm of 10 638.
TABLE 2.
Incidence of Cancer by Whether Infants Received Phototherapy, by Cancer Site
| Phototherapy (n = 178 017) | No Phototherapy (n = 4 966 832) | Unadjusted Risk Difference (per 100 000) | 95% CI (per 100 000) | P a | |||
|---|---|---|---|---|---|---|---|
| N b | Incidence (per 100 000) | N b | Incidence (per 100 000) | ||||
| All cancer | 58 | 32.6 | 1042 | 21.0 | 11.6 | 3.6 to 20.7 | .002 |
| Leukemia | 16 | 9.0 | 242 | 4.9 | 4.1 | 0.1 to 9.2 | .025 |
| Lymphoid leukemia | 6 | 3.4 | 133 | 2.7 | 0.7 | −1.7 to 4.2 | .49 |
| Myeloid leukemia | 10 | 5.6 | 103 | 2.1 | 3.5 | 0.4 to 7.8 | .006 |
| Brain or nervous system cancer | 7 | 3.9 | 176 | 3.5 | 0.4 | −2.2 to 4.1 | .69 |
| Eye or orbit cancer | 4 | 2.3 | 103 | 2.1 | 0.2 | −1.8 to 3.2 | .79 |
| Kidney cancer | 8 | 4.5 | 82 | 1.7 | 2.8 | 0.1 to 6.7 | .013 |
| Liver cancer | 2 | 1.1 | 81 | 1.6 | −0.5 | −1.9 to 2.0 | 1.0 |
| Soft tissue cancer | 1 | 0.6 | 69 | 1.4 | −0.8 | −1.9 to 1.3 | .52 |
| Other cancer | 26 | 14.6 | 426 | 8.6 | 6.0 | 0.8 to 12.4 | .013 |
Fisher’s exact test used.
Some subjects had >1 type of cancer.
TABLE 3.
RRs and Propensity-Adjusted ORs for the Association Between Phototherapy and Cancer, by Cancer Site
| N a | Unadjusted Analyses | Propensity-Adjusted Analysesb | |||||
|---|---|---|---|---|---|---|---|
| RR | 95% CI | P c | OR | 95% CI | P | ||
| All cancer | 1100 | 1.6 | 1.2 to 2.0 | .002 | 1.4 | 1.1 to 1.9 | .007 |
| Leukemia | 258 | 1.8 | 1.1 to 3.1 | .03 | 1.8 | 1.1 to 3.1 | .02 |
| Lymphoid leukemia | 139 | 1.3 | 0.6 to 2.9 | .49 | 1.3 | 0.6 to 2.9 | .55 |
| Myeloid leukemia | 113 | 2.7 | 1.4 to 5.2 | .006 | 2.6 | 1.3 to 5.0 | .005 |
| Brain or nervous system cancer | 183 | 1.1 | 0.5 to 2.4 | .69 | 1.0 | 0.5 to 2.1 | .97 |
| Eye or orbit cancer | 107 | 1.1 | 0.4 to 2.9 | .79 | 1.0 | 0.4 to 2.7 | .96 |
| Kidney cancer | 90 | 2.7 | 1.3 to 5.6 | .01 | 2.5 | 1.2 to 5.1 | .02 |
| Liver cancer | 83 | 0.7 | 0.2 to 2.8 | 1.0 | 0.6 | 0.2 to 2.5 | .5 |
| Soft tissue cancer | 70 | 0.4 | 0.1 to 2.9 | .52 | 0.4 | 0.1 to 2.7 | .33 |
| Other cancer | 452 | 1.7 | 1.1 to 2.5 | .01 | 1.6 | 1.1 to 2.4 | .02 |
Some subjects had >1 type of cancer.
Propensity-adjusted model included a propensity score for phototherapy that included the following variables: gender, birth wt, gestational age, large for gestational age, twin birth, birth by cesarean delivery, payer source, year of birth, maternal race, paternal race, maternal Hispanic ethnicity, maternal age, paternal age, maternal education, paternal education, Down syndrome, other chromosomal anomalies, and nonchromosomal congenital anomalies. The propensity score had an area under the receiver operating characteristic curve of 0.67.
Fisher’s exact test used.
Down syndrome was diagnosed in 7812 subjects (0.2% of the cohort). In the subgroup of infants with Down syndrome, 19% received phototherapy. Cancer was diagnosed in 18 infants with Down syndrome (lymphoid leukemia in 4, myeloid leukemia in 10, other leukemia in 2, brain or nervous system cancer in 1, and other cancer in 1). The marginal propensity-adjusted absolute risk increase for cancer after phototherapy was much higher for infants with Down syndrome (77.8/100 000; 95% CI, -1.2/100 000–156.8/100 000, number needed to harm of 1285) than for infants without Down syndrome (8.1/100 000, 95% CI 0.4/100 000–15.8/100 000, number needed to harm of 12 346). This difference was caused by the higher baseline risk of cancer in infants with Down syndrome; the propensity-adjusted ORs for cancer after phototherapy were similar: 1.2 for infants with Down syndrome and 1.4 for infants without Down syndrome (P for interaction = .79).
Discussion
In this study, we found that neonatal phototherapy was associated with an increased risk of cancer in the first year after birth; these associations persisted in adjusted analyses. Specifically, we found associations between phototherapy and overall cancer (adjusted odds ratio [aOR] 1.4), myeloid leukemia (aOR 2.6), kidney cancer (aOR 2.5), and other cancer (aOR 1.6). Based on our findings, 10 638 infants would need to be treated with neonatal phototherapy to cause 1 excess case of infantile cancer.
With epidemiologic studies, it is important to consider whether associations found are biologically plausible. The potential for phototherapy to increase the incidence of cancer is supported by evidence dating from the 1970s that blue light is mutagenic in vitro.11 Additionally, in vivo experiments on human newborns have demonstrated DNA damage, alterations in cytokine levels, and evidence of oxidative stress after treatment with phototherapy.12–22 This is concerning because all these conditions have been implicated in the pathogenesis of cancer.28,29 Although the clinical importance of these alterations remains to be elucidated, they present a potential mechanism whereby phototherapy could be related to childhood cancer.
The association between phototherapy and leukemia reported here has been previously suggested in other large epidemiologic studies. A population-based case–control study from the state of Washington found neonatal jaundice to be associated with an increased risk of leukemia (OR 1.5; 95% CI, 1.1–2.1).7 In a subgroup analysis, the risk of leukemia was higher in jaundiced infants who received phototherapy (OR 2.2; 95% CI, 1.0–4.9).7 A Swedish case–control study found increased odds of myeloid leukemia in infants with physiologic jaundice (OR 2.5; 95% CI, 1.2–5), which again was more pronounced in those treated with phototherapy (OR 7.5; 95% CI, 1.8–31.9).8 As in the current study, the Swedish group did not find an association between phototherapy and lymphoid leukemia (OR 1.0; 95% CI, 0.5–1.8).30
Two other studies did not find an association between phototherapy and cancer. A case–control study from the United Kingdom with 143 leukemia cases found no statistically significant association between phototherapy and leukemia (OR 0.5; 95% CI, 0.1–2.3); in this study, none of the 15 subjects with myeloid leukemia received phototherapy (OR 0; 95% CI, 0–11.7).10 A Danish study of 55 120 children with neonatal hyperbilirubinemia found no association between hyperbilirubinemia and overall cancer (standardized incidence ratio [SIR] 1.0; 95% CI, 0.8–1.3) or any subtype, including leukemia (SIR 1.2; 95% CI, 0.8–1.7) and kidney cancer (SIR 1.0; 95% CI, 0.4–2.2).9 The authors of that study estimated that 85% to 90% of the children with hyperbilirubinemia received phototherapy, but they were unable to specifically evaluate the association between phototherapy and cancer.
In this study, we also found an association between phototherapy and other cancer. The “other cancer” group included an excess of bone cancer, which is extremely rare in infancy, suggesting that this group consists, at least in part, of metastases and misclassified cancers. Nonetheless, we found a persistent association between phototherapy and these other cancers.
Our study has the advantage of a large cohort, with >5 million infants and 1100 cancer cases. The incidence of cancer in our study is similar to what is reported from US Cancer Registries,31 which suggests that our ascertainment of cancer cases was reasonably complete.
Because we used a large statewide data set, this study has a few limitations. We were unable to adjust for potential confounding variables not included in the data set. As an example, because serum bilirubin levels were not available, we cannot exclude the possibility that the association between phototherapy and cancer could be caused by elevated bilirubin levels (not captured by a discharge diagnosis of jaundice). The association between phototherapy and cancer persisted in analyses adjusted for a discharge diagnosis of jaundice, although a diagnosis of jaundice is admittedly a crude indicator of hyperbilirubinemia.
Because data on phototherapy intensity and duration were not available, we could not investigate the effect of phototherapy dosage. We did not have data on home phototherapy; we hypothesize that misclassification of subjects who received home phototherapy as unexposed would attenuate the observed association between phototherapy and cancer. Our ascertainment of phototherapy and of cancer diagnoses was probably incomplete, but it is unlikely that the completeness or accuracy of cancer coding differed between those who did and did not receive phototherapy or that the accuracy of phototherapy coding differed depending on subsequent risk of cancer. Such nondifferential misclassification, if present, would have biased our ORs toward 1. Also, the data set we used is limited to the first year after birth.
In a companion study, Newman et al3 looked at the association between neonatal phototherapy and childhood cancer in a cohort of almost 500 000 children born over a 17-year period in Kaiser Permanente Northern California hospitals. In that study, phototherapy was associated with an increased risk of overall cancer and nonlymphocytic leukemia in unadjusted models, but these associations were no longer statistically significant after researchers adjusted for potential confounders. These results could reflect the ability to adjust for additional potential confounders that were not available in our data set (eg, bilirubin levels). However, the 95% CI reported in that study did not exclude adjusted differences of the magnitude reported here. It is also possible that phototherapy causes only infantile or early childhood cancer; the Newman et al finding of increased early (<3 years) nonlymphocytic leukemia could support this hypothesis.
Although we found an increased risk of cancer in infants who received phototherapy, the absolute risk increase among those who received phototherapy was low. However, for infants with Down syndrome, the absolute risk increase was almost 10 times higher because of their higher baseline risk for cancer.
The possible increase in cancer risk from using phototherapy must be balanced against its benefit for lowering bilirubin levels. Phototherapy use has decreased rates of exchange transfusions,32 a procedure estimated to have 5% morbidity and up to 1.9% mortality risks.33 However, the number of infants with bilirubin levels near phototherapy treatment thresholds who need to be treated to prevent 1 from reaching exchange transfusion thresholds varies widely (from 10 to >3000, based on gender, gestational age, and age).34 Additionally, having bilirubin levels at or just above exchange transfusion thresholds is typically benign; it is not until bilirubin levels are >5 to 10 mg/dL above exchange transfusion thresholds that the risks of cerebral palsy and sensorineural hearing loss increase significantly.35–40 Therefore, especially in infants with bilirubin levels below recommended treatment thresholds and in infants with Down syndrome, the risks of phototherapy may exceed the benefits.
Conclusions
Phototherapy is a valuable treatment for infants with neonatal hyperbilirubinemia. Our results support the need for health care providers to consider phototherapy as they do most other treatments (ie, as having both benefits and potential risks) and to limit the use of phototherapy to infants for whom the benefit/risk balance is most likely to be favorable.
Acknowledgments
The authors thank Dr Beate Danielsen for her assistance on this project. Statistical analysis for this project was partially supported by the National Center for Advancing Translational Sciences, National Institutes of Health, through UCSF-CTSI grant UL1 TR000004. Its contents are solely the responsibility of the authors and do not necessarily represent the official views of the NIH.
Glossary
- aOR
adjusted odds ratio
- CI
confidence interval
- ICD-9
International Classification of Diseases, Ninth Revision
- OR
odds ratio
- RR
risk ratio
- SIR
standardized incidence ratio
- VS/PDD
Vital Statistics/Patient Discharge Dataset
Footnotes
Dr Wickremasinghe assisted with study design, obtained funding, obtained data, carried out the initial analyses, and drafted the initial manuscript; Dr Kuzniewicz assisted with study design and reviewed and revised the manuscript; Dr Grimes provided statistical consultation and reviewed and revised the manuscript; Dr McCulloch assisted with study design, provided statistical consultation, and reviewed and revised the manuscript; Dr Newman conceptualized and designed the study, assisted with obtaining funding and statistical analyses, and reviewed and revised the manuscript; and all authors approved the final manuscript as submitted.
FUNDING: Gerber Foundation Novice Research Award and American Academy of Pediatrics Marshall Klaus Perinatal Research Award to A.C.W.
Companions to this article can be found online at www.pediatrics.org/cgi/doi/10.1542/peds.2015-1354 and www.pediatrics.org/cgi/doi/10.1542/peds.2016-0983.
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