Pediatr Blood Cancer 2010;55:95–99
Infant Leukemia and Congenital Abnormalities: A Children’s Oncology Group Study
Kimberly J. Johnson, PhD,1Michelle A. Roesler, BS,1Amy M. Linabery, MS, MPH,1Joanne M. Hilden, MD,2
Stella M. Davies, MBBS, PhD,3and Julie A. Ross, PhD1,4*
Leukemia in infants is rare with an annual incidence rate of
?40 cases per million children <1 year of age in the United States
. Infant leukemia differs from that of older children in
several respects including a lower 5-year survival rate (?50% vs.
67–87%)  and translocations involving the mixed lineage
leukemia (MLL) gene in a majority of cases [3,4]. These
observations suggest that the etiology of infant leukemia is distinct
from leukemias arising in older children.
Congenital abnormalities (CAs) and chromosomal syndromes
have frequently been associated with childhood cancers and
leukemia in particular. Most notably, Down syndrome is firmly
suggested that CAs may be linked to leukemia including cleft
lip/palate , rib abnormalities [7–10], other chromosomal
abnormalities , pancreas–digestive tract abnormalities ,
large or multiple birthmarks, and heart defects [12,13].Results
of one study suggest that children born with CAs are more likely to
be diagnosed with cancer during infancy than those without a CA
. The association between CAs and leukemia in infants,
however, has not been specifically evaluated.
Using data from a Children’s Oncology Group (COG) case–
and maternally reported CAs. Results from this study may give
insights into its pathogenesis.
Institutional Review Boards at the University of Minnesota and
each case’s COG institution approved the study protocol. Details of
the case–control study design and data collection have been
previously described [14–16] and are summarized below.
This study was conducted in two phases. Acute lymphoblastic
leukemia (ALL) and acute myeloid leukemia (AML) cases
diagnosed from birth to age 1 year during January 1, 1996 to
October 13, 2002 (phase one) and January 1, 2003 to December 31,
2006 (phase two) were identified through participating COG
from the study if they had Down syndrome, their mother did not
speak English or Spanish (Spanish language included in phase two
only), orif their biologicalmotherwas not available foratelephone
interview. Leukemia subtype and MLL-rearrangement status were
verified as previously described [15,16].
Two hundred forty of 348 potential phase one cases completed
participating COG institutions were enrolled in the study. Two
hundred three (59%) mothers completed interviews, while the
remainder did not due to maternal refusal (22%), inability to locate
the mother (11%),physician refusal (6%), and institutional failure to
(May 1999 to October 2002) were sampled from the population
 with minor modifications and were frequency matched to cases
Background. Leukemia in infants is rare and has not been well
studied apart from leukemia in older children. Differences in survival
and the molecular characteristics of leukemia in infants versus older
children suggest a distinct etiology, likely involving prenatal factors.
Procedure. We examined the association between eight categories
of maternally reported congenital abnormalities (CAs) (cleft lip or
palate, spina bifida or other spinal defect, large or multiple
birthmarks, other chromosomal abnormalities, small head or micro-
cephaly, rib abnormalities, urogenital abnormalities, and other) and
infant leukemia in a case–control study. The study included
443 cases diagnosed at <1 year of age at a Children’s Oncology
Group Institution in the United States or Canada from 1996 to 2006
and 324 controls. Controlswere recruited from the cases’ geographic
area either by random digit dialing (1999–2002) or through birth
certificates (2003–2008) and were frequency-matched to cases on
birth year. Odds ratios (ORs) and 95% confidence intervals (CIs)
were calculated by unconditional logistic regression after adjustment
for birth year and a measure of follow-up time to account for
differences in the CA observation period. Results. No statistically
significant associations were observed between infant leukemia and
any CA (OR¼1.2; 95% CI: 0.8–1.9), birthmarks (OR¼1.4; 95% CI:
0.7–2.5), urogenital abnormalities (OR¼0.7; 95% CI: 0.2–2.0), or
other CA (OR¼1.4; 95% CI: 0.7–2.8). Results were similar for acute
lymphoblastic and myeloid leukemia cases. Fewer than five
subjects were in the remaining CA categories precluding analysis.
Conclusions. Overall, we did not find evidence to support an
association between CAs and infant leukemia. Pediatr Blood Cancer
? 2010 Wiley-Liss, Inc.
birth defects; congenital abnormalities; infant leukemia; risk factors
? 2010 Wiley-Liss, Inc.
Published online 27 March 2010 in Wiley InterScience
Additional Supporting Information may be found in the online version
of this article.
2Department of Pediatric Hematology/Oncology, Peyton Manning
Children’s Hospital at St. Vincent, Indianapolis, Indiana;3Children’s
Hospital Medical Center, Cincinnati, Ohio;4Masonic Cancer Center,
University of Minnesota, Minneapolis, Minnesota
Conflict of interest: Nothing to declare.
Grant sponsor: NIH; Grant numbers: R01 CA79940, R01 CA067263,
U10 CA98413, U10 CA98543, T32 CA099936; Grant sponsor:
Children’s Cancer Research Fund, Minneapolis, MN.
*Correspondence to: Julie A. Ross, Professor and Director, Division of
Pediatrics, Children’s Cancer Research Fund Chair in Molecular
Epidemiology, MMC 422, 420 Delaware St. S.E., Minneapolis, MN
55455. E-mail: firstname.lastname@example.org
Received 24 November 2009; Accepted 1 February 2010
on birth year. The field response rate of 59% for phase one controls
was calculated according to the method by Slattery et al.  as
previously detailed . Phase two controls (October 2003 to
March 2008) were ascertained through state birth registries. Sixteen
states were initially selected for control recruitment because of their
ability to release birth records and because a substantial number of
phase one cases were diagnosed inthese states; 15 of these approved
this study and provided lists of potential controls randomly selected
from their birthregistries. Controls were frequency matchedtocases
on birth year and region of residence based on the phase one case
distribution with a recruitment goal of 150 additional controls. Two
hundred seventy potential controls were approached with a mailed
letter, of which 150 were located. Of these, 21 passively refused,
55 actively refused, and 3 were ineligible resulting in 71 completed
controls were similar to one another based on maternal and infant
one control group.
Information on infant birth characteristics and maternal
characteristics was collected through a structured telephone inter-
view. Covariates considered in this analysis included infant sex,
birth weight, gestational age, parental age, parental education,
parental race, and annual household income. Information on the
presence of Down syndrome and eight CA categories (cleft lip
or palate; spina bifida or other spinal defect; large or multiple
birthmarks; other chromosomal abnormalities; unusually small
organ abnormalities (urogenital abnormality); any other birth
defect) in case and control subjects and their siblings was collected
during the mother’s telephone interview. If mothers reported an
‘‘other’’ CA in their offspring, they were asked to specify the type.
All statistical analyses were conducted using SAS version 9.1
(Cary, NC). Statistical differences in the frequency or means
of sociodemographic and infant characteristics between cases
and controls were assessed using Fisher’s exact test and one-way
analysis of variance for categorical and continuous variables,
respectively. We used unconditional logistic regression to examine
their siblings after adjusting for birth year. Since some CAs are not
readily observeduntil after the first year , we also adjusted fora
measure of follow-up time in models for associations between CAs
and infant leukemia in index children. Follow-up time was defined
as the interval between the birth date and interview or death date,
whichever came first and was dichotomized at the median (based
on both case and control subjects). Potential confounders of
the association between CAs and infant leukemia overall were
examined using a forward selection process  where each
potential confounder was added one at a time to models that
included birth year to determine their effect on the parameter
estimate. If a potential confounder materially changed the
interpretation of the result, it was retained in the model. All
statistical tests were two-sided.
Approximately 60% and 39% of infant cases had a diagnosis of
ALL or AML, respectively, with the remaining cases having
diagnoses of biphenotypic or acute unknown leukemias. Among
cases with known MLL-rearrangement status, over half were
MLL-positive with a ratio of ?2:1 and 1:1 for ALL and AML
MLL-positive to negative cases, respectively. The mean diagnosis
age was ?6 months overall with MLL-positive cases tending to be
diagnosed earlier than other cases. There were more females than
males among MLL-positive cases with the reverse finding among
MLL-negative and unknown cases (Table I).
The mean gestational age and birth weight was similar between
cases and controls, while cases were more likely than controls to be
first born (Table II). Case mothers were slightly younger, and more
likely to report household incomes of ?$30,000, being unmarried,
and non-White. Notably, over twice as many case than control
mothers reported being of Hispanic ethnicity. A similar percentage
of case and control mothers reported having college degrees.
Logistic regression results for associations between infant
leukemia and CAs with ?5 case and control subjects in index
children are provided in Table III. There were no significant
associations between infant leukemia overall and any CA (odds
ratio (OR)¼1.2; 95% confidence interval (CI); 0.8–1.9), large or
multiple birthmarks (OR¼1.3; 95%CI: 0.7–2.4), urogenital
system abnormalities (OR¼0.7; 95%CI: 0.2–2.0), or other CAs
(OR¼1.4; 95% CI: 0.7–2.8) in models that adjusted for birth year
and follow-up time category or in those that also included other
covariates shown in Table II (data not shown). Results were similar
when analyzed for ALL and AML infant leukemia subtypes
specifically. We observed no significant effect modification by sex
with the exception of a marginally significant interaction (P for
interaction¼0.1) between large or multiple birthmarks and infant
Pediatr Blood Cancer DOI 10.1002/pbc
TABLE I. Characteristics of Infant Leukemia Diagnoses
MLL-positive (n¼228) MLL-negative (n¼146) MLL-unknown (n¼69)Total (n¼443)
Leukemia subtype, no. (%)
Diagnosis age category, no. (%)
Gender, no. (%)
96Johnson et al.
95% CI: 0.9–5.1). Risks of similar magnitude in association with
birthmarks were apparent in females for ALL (OR¼2.4; 95% CI:
0.9–6.4) and AML (OR¼2.7; 95% CI: 0.9–7.9) (Supplementary
CAs varied by MLL-rearrangement status. MLL-positive cases had
non-significant increased risks of leukemia in association with
birthmarks (OR¼1.5; 95% CI: 0.7–3.2) and other CAs (OR¼2.0;
95% CI: 0.9–4.5; Supplementary Table IV). Similar percentages of
and 11.8%, respectively). We found significant effect modification
by sex of the association between birthmarks butnot other CAs and
MLL-positiveleukemia (P for interaction¼0.05) with an increased
risk in females (ORfemales¼2.9; 95% CI: 1.0–7.9) but not males
(ORmales¼0.7; 95% CI: 0.2–2.1; Supplementary Table V).
Finally, we examined whether CAs in the siblings were
associated with infant leukemia. No significant associations were
found between sibling CAs and infant leukemia for any CA
(OR¼1.1; 95% CI: 0.7–1.6), large or multiple birthmarks
(OR¼0.9; 95% CI: 0.5–1.7), urogenital system abnormalities
1.9; Supplementary Table VI).
In this study, we did not find statistical evidence for an
association between maternally reported CAs and infant leukemia.
Previous studies have suggested that severaltypes of CAs confer an
increased risk for leukemia. However, not all previous studies have
excluded individuals with Down syndrome from their analyses
[11,21]. In studies that have excluded these individuals, non-
significant associations have generally been reported between
leukemia (AML, ALL, or total) and any congenital anomaly
[13,22–29], major congenital anomalies , and heart defects
Pediatr Blood Cancer DOI 10.1002/pbc
TABLE II. Birth Characteristics and Parental Sociodemographics of Cases and Controls
Gestational age category, no. (%)
Birth weight category (g), no. (%)
First-born, no. (%)
Maternal age category, no. (%)
Household income category (US dollars)b, no. (%)
Marital status, no. (%)
Maternal race, no. (%)
White, not of Hispanic origin
African-American or Black, not of Hispanic origin
Native American Indian or Alaskan Native
Asian, Asian-American, or Pacific Islander
Maternal educationc, no. (%)
190 (42.9) 0.07
aP-values calculated by Fisher’s exact test;bThree cases and four controls had missing data on household
income;cOne case and one control had missing data on maternal education.
Infant Leukemia and Congenital Abnormalities97
[13,22], while significant associations have been reported for
birthmarks (ALL and AML) [12,31], heart abnormalities (ALL)
, and pancreas or digestive tract abnormalities (ALL) . Our
consistent with that of most previous studies that have excluded
Our finding of a twofold increased risk of infant leukemia in
inlight of reports by four previous studies (two of childhood cancer
and two of childhood leukemia) that have reported significant
positive associations for birthmarks [12,31–33]. Mertens et al. 
reported significant increased risks for large or multiple birthmarks
overall in the index child but not their siblings for both ALL
(OR¼1.3; 95% CI: 1.1–1.7) and AML (OR¼1.9; 95% CI: 1.2–
3.1). Roganovic et al.  reported significant excesses of
pigmented nevi and cafe ´-au-lait spots in children with hemato-
reported increased frequencies of birthmarks and port-wine stains,
respectively, in childhood cancer cases compared to non-cases.
these findings is unclear as none of these studies has included
cases that were reported to have neurofibromatosis type 1 that is
characterized by the presence of cafe ´-au-lait spots and an increased
risk of leukemia .
Intriguingly, four studies have reported significant positive
associations between rib abnormalities and leukemia [7–10].
However, only two of these studies reported excluding children
with Down syndrome, which has previously been noted to be
associated with an excess of rib abnormalities . Two controls
expected frequency [8–10], that is likely due to under-reporting
since mothers may not be aware of rib abnormalities in the absence
of their child having undergone a chest X-ray.
One previous study has suggested that children born with a CA
without a CA  implicating the infant period as being the most
vulnerable to development of malignancy. The risk for leukemia
was mainly due to chromosomal anomalies (including Down
syndrome) and not other defects, which is consistent with our
This study has strengths and limitations. Our study is the largest
all data were collected systematically by trained interviewers. A
limitation of our study is that CAs were ascertained by maternal
sizes. Forexample, wehad 80%powertodetect anincreased OR of
1.7 for the overall association between CAs and infant leukemia.
Recall or reporting bias could have affected our results if the
bias could result if mothers over-reported CAs in case children.
However, this seems unlikely to have occurred since associations
bias may also have affected our results since our controls were
different on several respects from cases and the general population
as previously reported . However, adjustment for factors that
differed between cases and controls did not materially change the
In conclusion, this study does not provide evidence for a strong
association between CAs and infant leukemia.
This study was supported by NIH grants R01 CA79940, R01
CA067263, U10 CA98413, U10 CA98543, and T32 CA099936,
and the Children’s Cancer Research Fund, Minneapolis, MN.
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TABLE III. Associations Between Infant Leukemia and Maternally Reported Congenital Abnormalities Overall and by Leukemia
abnormality categoryControls (no.)
95% CI Cases (no.) ORa
95% CICases (no.)ORa
Large or multiple birthmarks
Urogenital system abnormality
0.2–3.2 0.70.8 0.7
0.7–3.8 1.4 1.5 1.6
aAdjusted for birth year and follow-up time category dichotomized at the median (27, 32, and 34 months for total, AML, and ALL analyses,
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