ARTICLE

Birth Weight and Risk for Childhood Leukemia in Denmark, Sweden, Norway, and Iceland

Lisa Lyngsie Hjalgrim, Klaus Rostgaard, Henrik Hjalgrim, Tine Westergaard, Harald Thomassen, Erik Forestier, Göran Gustafsson, Jon Kristinsson, Mads Melbye, Kjeld Schmiegelow

Affiliations of authors: Department of Epidemiology Research, Danish Epidemiology Science Center, Statens Serum Institut, Copenhagen, Denmark (LLH, KR, HH, TW, MM); Department of Pediatrics, Haukeland University Hospital, Bergen, Norway (HT); Department of Clinical Science, Pediatrics, University of Umeaa, Umeaa, Sweden (EF); Astrid Lindgren’s Children’s Hospital, Paediatric Cancer Research Unit, Stockholm, Sweden (GG); Children’s Hospital, University Hospital, Reykjavik, Iceland (JK); Department of Pediatrics, University Hospital, H:S. Rigshospitalet, Copenhagen (KS)

Correspondence to: Lisa Lyngsie Hjalgrim, MD, Department of Epidemiology Research, Statens Serum Institut, Artillerivej 5, 2300 Copenhagen S, Denmark (e-mail: lih{at}ssi.dk)


    ABSTRACT
 Top
 Notes
 Abstract
 Introduction
 Patients and Methods
 Results
 Discussion
 References
 
Background: Compelling evidence suggests that childhood leukemia often originates in utero. Birth weight is one of the few pregnancy-related risk factors that has been associated with leukemia risk, but the association has remained poorly characterized. We conducted a population-based case–control study in Denmark, Sweden, Norway, and Iceland to investigate the association between birth weight (and other birth characteristics) and the risk of childhood leukemia. Methods: Overall, 1905 children (aged 0–14 years) with acute lymphoblastic leukemia (ALL) and 299 children with acute myeloid leukemia (AML) diagnosed between January 1, 1984, and December 31, 1999, were identified in the Nordic Society of Paediatric Haematology and Oncology acute leukemia database. Each case patient was matched to five population control subjects (n = 10745) on nationality, age, and sex. All live-born siblings of case patients (n = 3812) and control subjects (n = 17 937) were also identified in population registers. Information on birth weight and gestational age at birth was ascertained from the national Medical Birth Registers. The association between various birth characteristics and leukemia risk was assessed by conditional logistic regression. All statistical tests were two-sided. Results: Risk of ALL overall was statistically significantly associated with birth weight (odds ratio [OR] = 1.26 per 1-kg increase in birth weight, 95% confidence interval [CI] = 1.13 to 1.41). The association was similar for B- and T-lineage ALL and across all diagnostic ages (0–14 years). However, children with ALL did not weigh more at birth than their siblings. Statistically significantly reduced risks of B-precursor ALL were observed with increasing position in the birth order (OR = 0.90 per position increase, 95% CI = 0.84 to 0.96) and increasing gestational age (OR = 0.87 per 2-week increase in gestational age, 95% CI = 0.81 to 0.94). Risk of AML did not vary monotonically with birth weight, and low birth weight (<1500 g [i.e., 3.3 pounds]) was associated with the highest risk. Conclusion: Our results are compatible with the hypothesis that a high birth weight is associated with an increased risk of ALL.



    INTRODUCTION
 Top
 Notes
 Abstract
 Introduction
 Patients and Methods
 Results
 Discussion
 References
 
The development of both acute lymphoblastic leukemia (ALL) and acute myeloid leukemia (AML) in childhood is frequently initiated in utero (16). However, the pathogenesis of childhood ALL and AML is poorly understood, and few prenatal risk factors have been identified. One possible risk factor is birth weight. A recent meta-analysis of more than 10000 patients with leukemia concluded that a birth weight of more than 4000 g (i.e., 8.8 pounds) was associated with an increased risk of ALL (odds ratio [OR] = 1.26, 95% confidence interval [CI] = 1.17 to 1.37), compared with birth weights of less than 4000 g, and that the association between birth weight and leukemia appeared to follow a log-linear dose–response relationship (7). A more detailed characterization of the association between birth weight and leukemia is warranted for determining the possible underlying biologic mechanisms. Specifically, because childhood leukemia is a group of biologically and clinically distinct diseases (each with its own characteristic epidemiology and genetic alterations), the natural history of each disease, including the possible association with birth weight, may also be distinct (8). Thus, it is unclear whether the association with birth weight applies to all or only certain leukemia subtypes (911) or whether birth weight is associated with age at diagnosis (1218). It is also unclear whether high birth weight is a trait unique to children with leukemia or a trait shared with the siblings of children with leukemia (19).

To further characterize the association between birth weight and leukemia, we conducted a nationwide register-based case–control study in Denmark, Sweden, Norway, and Iceland to investigate birth weight and other birth characteristics as risk factors for childhood leukemia, with special emphasis on leukemia subtypes. Moreover, we compared birth weight patterns of case patients with their siblings and patterns of case siblings with control siblings.


    PATIENTS AND METHODS
 Top
 Notes
 Abstract
 Introduction
 Patients and Methods
 Results
 Discussion
 References
 
Study Subjects

Case children were identified in the population-based Nordic Society of Paediatric Haematology and Oncology (NOPHO) acute leukemia database. This database was established by NOPHO in 1981 to promote and monitor uniform diagnostic, treatment, and clinical follow-up procedures. The database contains information on all children diagnosed with ALL (since July 1981) and AML (since July 1984) in the Nordic population of approximately 5 million 0- to 14-year-old children (20,21). Case patients were children (aged 0–14 years) diagnosed with ALL from January 1, 1984, through December 31, 1999, or with AML from July 1, 1984, through December 31, 1999, in Denmark, Sweden, Norway, or Iceland. This study was approved by the national data protection boards in all four participating countries.

Each patient is registered with a unique personal identification number (which includes information on date of birth and sex). Such personal identification numbers have been assigned to all residents of the Nordic countries since the mid-1960s and are used by the civil registration systems to continuously monitor vital status of the populations (22). The number is also used by health registers, thus making register linkages possible. For each patient, information was extracted from the NOPHO acute leukemia database on nationality, Down syndrome (yes or no), and data pertaining to the leukemia, including information on date of diagnosis, histologic subtype, immunophenotype, and karyotype. Specifically, case patients with ALL were classified according to their immunophenotype, i.e., as B-precursor ALL, mature B-cell ALL, T-cell ALL, or ALL not otherwise specified, if registry information did not allow proper classification. Information on karyotype has routinely been registered in the NOPHO acute leukemia database within the last 10 years (23), except for t(12;21), which has been registered for a substantial number of patients, but only within the last 5 years. Case patients with AML were grouped according to the modified French, American, and British classification system (FAB classification M0–M7) (24).

Using the personal identification number of each case patient, we also identified five population control subjects individually matched to case patients on sex, birth month, and birth year, who were alive on the date of diagnosis and were listed in the Civil Registration System (Denmark) (25) and in the nationwide Medical Birth Registers for Sweden, Norway, and Iceland (2630). Through the Civil Registration System in Denmark (25), the Medical Birth Registers in Norway and Iceland (30,31), and the Multi-Generation Register in Sweden (32), we identified all parents and live-born siblings (defined as children with the same mother as the index child) of case patients and control subjects. Birth dates on siblings were used to create complete birth order data sets for both case patients and control subjects. Information on birth weight and gestational age at birth for case patients, control subjects, and their respective siblings was extracted from the population-based Medical Birth Registers in Denmark (established in 1973), Sweden (established in 1973), Norway (established in 1967), and Iceland (established in 1972). These registers contain mandatorily reported data from midwives on all births in the respective countries (2628,30,31).

We identified 2041 children with ALL and 379 children with AML in the NOPHO acute leukemia database. Ninety-five percent of the case patients were identified in the medical birth registers, resulting in 1941 case patients with ALL and 360 case patients with AML. After the exclusion of case patients with Down syndrome (n = 35 for ALL, and n = 60 for AML) and their respective control subjects, and the exclusion of case patients and control subjects who, for other reasons (e.g., foreign nationality or uncertainty as to mother’s identity), did not have relevant information recorded in the Medical Birth Registers, the analysis included a total of 1905 children with ALL and 299 children with AML, a total of 9296 and 1449 matched control subjects, respectively, and a total of 3812 siblings of case patients and 17 937 siblings of control subjects.

Statistical Analyses

We studied whether birth weight, gestational age, birth order, maternal age, and paternal age were associated with the risk of ALL overall, B-precursor ALL, T-cell ALL, t(12;21)-positive ALL, hyperdiploidy-positive ALL, and AML. No subtype analyses were conducted for mature B-cell ALL (n = 33) and unspecified ALL (n = 49) because of the small number of cases. Gestational age at birth was measured in completed pregnancy weeks, which for the vast majority of the children was based on date for last menstrual period. Birth order was defined as one plus the previous number of live children born to the same mother. Twins were assigned the same position in birth order as if they had been singletons. Maternal age and paternal age were the decimal age of the mother and father, respectively, when the child in question was born.

We used conditional logistic regression to estimate odds ratios with the PHREG procedure in SAS (33), and thus implicitly always adjusted for the matching factors of country, sex, and date of birth. P values and 95% confidence intervals were based on two-sided Wald tests. In univariate analyses of leukemia risk, we treated birth weight, gestational age, and birth order as categorical variables to give a detailed impression of the effects of these variables on leukemia risk. In other analyses, we modeled birth weight, gestational age, birth order, maternal age, and paternal age as a trend. Model reduction tests in both univariate and multivariate analyses showed all exposure effects to be adequately described by trends. This was the case whether the starting model modeled these effects categorically or through smoothing splines (34). We also analyzed whether the association with birth weight was modified by age at diagnosis, sex, country, or gestational age. Specifically for age at diagnosis, we tested whether additional interaction terms in the model involving birth weight and age reached statistical significance. These interaction terms assumed that the birth weight trend varied as a first-order or second-order polynomial of age or, alternatively, was specific for an age category (0, 1–5, 6–9, or ≥10 years). Moreover, associations of risk with birth weight, gestational age, or birth order were compared between ALL and AML and between B-precursor ALL and T-cell ALL by the data duplication method described by Lunn and McNeil (35). The tests for homogeneity of effects for different outcomes were performed as chi-square–Wald tests.

We compared birth weights between case patients and their siblings and between case siblings and control siblings, adjusting for other birth characteristics. In the birth weight analysis, we included only those siblings whose parents were identical to those of the corresponding case patient or control subject. We excluded index children (case patients and control subjects) for any of the following reasons: 1) if they had missing information on father’s birth date, 2) if an older sibling had missing information on the father’s birth date, or 3) if the father’s birth date did not match that of the father of an older sibling. If a younger sibling of the index child was a half-brother or half-sister, only the sibling was excluded. All twin pairs (among case patients, control subjects, and their siblings) were excluded from this analysis. The total number of children available for this analysis was then 1598 ALL case patients and their 7621 control subjects, 248 AML case patients and their 1183 control subjects, including 2491 case siblings and the their 11 622 control siblings.

We assumed that individual birth weights for case siblings were composed of the following components: 1) a systematic effect of sex-specific birth order position (0, 1, 2, 3, or ≥4), gestational age, sex, country, mother’s age, and birth date; 2) a systematic effect of status as a case patient, status as a case sibling born more than 9 months after leukemia diagnosis (anticipating maternal behavioral changes affecting birth weight), and status as a case sibling; 3) a random familial component (shared by all siblings with common parents); and 4) random variation between children. The effects of component 1 were estimated among control siblings with ordinary linear regression and entered as an offset. Birth order, country, and sex were treated as factors, i.e., each possible value of the predictor was assigned its own parameter. The random components (3 and 4) were assumed to be independently and normally distributed, with a mean of zero, and to be identically and independently distributed between children and families (siblings with common parents), respectively. The model was fitted by use of the MIXED procedure in SAS (33). For technical reasons, we modeled the difference between birth weight and the systematic effects in component 1 by interpreting the intercept as the effect of case sibling status.


    RESULTS
 Top
 Notes
 Abstract
 Introduction
 Patients and Methods
 Results
 Discussion
 References
 
Distribution of Leukemia Case Patients

The sex and subtype distribution of the 1905 ALL case patients and 299 AML case patients from Denmark, Sweden, Norway, and Iceland is shown in Table 1. Of the 1638 case patients with B-precursor ALL, 101 were t(12;21)-positive and 319 were hyperdiploidy-positive.


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Table 1. Distribution of patients with leukemia by country and subtype

 
Birth Weight

The birth weight of 99% of the case patients (with ALL or AML) and control subjects was recorded in the Medical Birth Registers. In unadjusted dichotomous analyses, a birth weight of 4000 g or more was associated with a statistically significantly higher risk of ALL than a birth weight of less than 4000 g (OR for ALL overall = 1.18, 95% CI = 1.03 to 1.34; OR for B-precursor ALL = 1.16, 95% CI = 1.00 to 1.34; and OR for T-cell ALL = 1.51, 95% CI = 1.01 to 2.26). When we used birth weight as a continuous variable and adjusted for gestational age, birth order, and parental age, risk of ALL and ALL subtypes was positively associated with birth weight (adjusted OR for B-precursor ALL = 1.30 per 1-kg increase in birth weight, 95% CI = 1.16 to 1.47; and the adjusted OR for T-cell ALL = 1.14 per 1-kg increase in birth weight, 95% CI = 0.80 to 1.62) (P for homogeneity = .46) (Table 2), as was that for hyperdiploidy (adjusted OR for positive ALL = 1.39 per 1-kg increase in birth weight, 95% CI = 1.05 to 1.83) and that for t(12;21) (adjusted OR for positive ALL = 1.43 per 1-kg increase in birth weight, 95% CI = 0.90 to 2.20). The association between birth weight and leukemia risk did not vary by age (whether categorized [as shown in Table 3 ] or continuous [data not shown]), sex, country (data not shown), or gestational age (data not shown).


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Table 2. Adjusted odds ratios (ORs) and 95% confidence intervals (CIs) for acute lymphoblastic leukemia (ALL) overall, ALL subtypes, and acute myeloid leukemia (AML) by birth characteristic*

 

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Table 3. Adjusted odds ratios (ORs) and 95% confidence intervals (CIs) for acute lymphoblastic leukemia (ALL) and acute myeloid leukemia (AML), stratified by age*

 
Unadjusted dichotomous analyses showed that birth weights of 4000 g or more were associated with a statistically nonsignificantly greater risk of AML than birth weights of less than 4000 g (OR = 1.19, 95% CI = 0.85 to 1.68). When adjusted for birth order, gestational age, and parental age, the risk of AML (OR per 1-kg increase in birth weight = 1.09, 95% CI = 0.82 to 1.45) was not statistically significantly different from that observed for ALL overall (adjusted OR for ALL overall = 1.26 per 1-kg increase in birth weight, 95% CI = 1.13 to 1.41; P for homogeneity = .33). As for ALL, the association between birth weight and AML risk did not vary by country, sex (data not shown), or age group (Table 3). However, although the association between birth weight and risk of AML was statistically consistent with the a priori hypothesized linear trends (data not shown), alternative association patterns might also be considered. Specifically, the data for AML suggested a U-shaped association between birth weight and AML risk (Table 2). Thus, the risk of AML associated with birth weights of less than 2500 g (adjusted OR for AML = 1.56, 95% CI = 0.71 to 3.44) and birth weights of 4000 g or more (adjusted OR for AML = 1.17, 95% CI = 0.83 to 1.65) was greater than that associated with birth weights of 2500–3999 g.

Gestational Age

Gestational age at birth was recorded for 97% of the case patients and control subjects. After adjusting for birth weight, birth order, and parental age, gestational age was inversely associated with risk of B-precursor ALL (OR = 0.87 per 2-week increase in gestational age, 95% CI = 0.81 to 0.94) (Table 2). Gestational age was not statistically significantly associated with the risk of T-cell ALL or AML (Table 2). The magnitudes of the association of gestational age with the risks of T-cell ALL and B-precursor ALL were statistically significantly different (P for homogeneity = .04), but the magnitudes of the association with the risks of ALL and AML were not statistically different (P for homogeneity = .10). The association of gestational age with risk of B-precursor ALL did not vary by country or age group (data not shown) but was slightly more pronounced for girls than boys (P for homogeneity = .04).

Birth Order

An increasingly later position in the birth order was associated with a statistically significantly decreased risk of ALL overall (OR = 0.91 per one position increase in birth order, 95% CI = 0.85 to 0.97) and B-precursor ALL (OR = 0.90 per one position increase in birth order, 95% CI = 0.84 to 0.96) (Table 2). Increasing birth order position was associated with a decreased risk of t(12;21)-positive ALL (OR per one position increase in birth order = 0.95, 95% CI = 0.71 to 1.27) and of hyperdiploidy ALL (OR per one position increase in birth order = 0.95, 95% CI = 0.80 to 1.10), although neither association was statistically significant. Birth order was not associated with the risk of T-cell ALL or risk of AML (Table 2). However, the association between birth order and the risk of B-precursor ALL and between birth order and risk of T-cell ALL did not differ statistically significantly (P for homogeneity = .20) and neither did the associations between birth order and the risk of ALL overall and between birth order and the risk of AML (P for homogeneity = .85).

Parental Age

We found no association between maternal age or paternal age and the risk of any leukemia subtype (ALL, B-precursor ALL, T-cell ALL, or AML) (data not shown).

Birth Weight of Case Siblings and Control Siblings

We investigated whether case patients were heavier at birth than their siblings and whether case siblings were heavier at birth than control siblings. In a model predicting birth weight for siblings of case patients and for siblings of control subjects, we found that ALL case patients did not weigh more at birth than their siblings (difference of the mean weights = 3 g, 95% CI = –22 to 27 g), but that ALL case siblings weighed 47 g (95% CI = 26 to 70 g) more than control siblings. AML case patients weighed 34 g (95% CI = –34 to 102 g) more than their siblings, whereas AML case siblings weighed 7 g (95% CI = –69 to 56 g) less than control siblings. All analyses were adjusted for sex-specific birth order, sex, country, gestational age, maternal age, and birth date.


    DISCUSSION
 Top
 Notes
 Abstract
 Introduction
 Patients and Methods
 Results
 Discussion
 References
 
Using a large dataset from nationwide population-based registers in Denmark, Sweden, Norway, and Iceland, we observed a statistically significant 26% increase in risk of ALL per 1-kg increase in birth weight. Although this observation is consistent with previous findings (7), use of the detailed information in our dataset allowed us to characterize, to our knowledge for the first time, the association between birth weight and the risk of childhood leukemia with respect to disease subtype and familial determinants.

First, our data indicate that the association with birth weight was not restricted to any one subtype of ALL but applied to childhood ALL in general. Birth weight was associated with the risk of B-precursor ALL and the risk of T-cell ALL; in B-precursor ALL subset analyses, birth weight was associated with risks of t(12;21)- and hyperdiplody-positive subtypes. These uniformly increased risks of different ALL subtypes indicate that the determinants of the elevated birth weights in leukemic children are not risk factors for a specific type of ALL but, rather, that a more universal phenomenon underlies the observed association.

Second, it was not known whether a higher birth weight was unique to the children with leukemia or to their families. Such information is, however, relevant to the determination of possible biologic mechanisms for the association between birth weight and leukemia risk. Using information on birth weights of siblings to children with ALL and to control subjects, we found that children in families afflicted by childhood ALL weighed more at birth (47 g more) than children from families not afflicted with this disease but that the birth weight of children with leukemia did not differ from that of their siblings. Our findings therefore contrast with those of a smaller previous study (19) of 72 sibships that reported higher birth weights for children with leukemia than for their siblings.

Various mechanisms have been suggested to support the association between birth weight and leukemia risk. Specifically, the level of insulin-like growth factor I is associated with birth weight and may play a role in the development of childhood leukemia through the induction of proliferative stress on progenitor cells or pre-leukemic cells (i.e., an increase in the number of cell divisions) (36). Another hypothesis suggests that, because there is an association between birth weight and bone marrow volume (number of bone marrow cells) (37,38), children with a higher birth weight have more cells at risk of malignant transformation and are thus at a greater risk of leukemia (39,40). It should be noted that these two hypotheses are not mutually exclusive. In fact, a third hypothesis is that pre-leukemic cells secrete growth factors that increase birth weight (36,41). Our findings clearly provide evidence of a strong familial component in the higher birth weight of children with ALL, and this observation is not easily compatible with the third hypothesis of reverse causality, namely that pre-leukemic cells stimulate growth and cause high birth weight in children with leukemia. However, our findings do not allow us to rule out either of the first two alternative hypotheses. Thus, our results are consistent with birth weight being associated with the proliferative stress on progenitor cells and/or birth weight being associated with the total number of cells at risk of malignant transformation.

The development of childhood leukemia may be a multistep process that involves the acquisition of both pre- and postnatal genetic aberrations (26,42), and more children (100-fold more) may be born with leukemia-associated genetic aberrations than will ultimately develop ALL (43). This evidence strongly suggests that postnatal risk factors are critical to the development of ALL. Another interpretation for the association between birth weight and ALL risk may be that, rather than determining the risk of prenatal initiation of ALL development as a factor of the number of normal (susceptible) cells, birth weight modifies ALL risk as a factor of the number of leukemia-initiated cells that are at risk of completing the ALL developmental process. In contrast to previous studies (13,1618), we found no association between birth weight and age at leukemia diagnosis. This result indicates that postnatal ALL risk factors may be independently distributed by birth weight.

Whether a similar association with birth weight applies to AML risk is less clear from our results. Although the observed trend estimate (OR = 1.09, 95% CI = 0.82 to 1.45 per 1-kg increase in birth weight) could be assumed to be statistically identical to the one observed for ALL overall (OR = 1.26, 95% CI = 1.13 to 1.41 per 1-kg increase in birth weight), the data are also suggestive of a U-shaped association between birth weight and AML risk, with children with birth weights at both ends of the spectrum having higher risk of AML than normal birth weight children. Other studies examining the association of high birth weight with AML risk have also arrived at conflicting results (17,18,4448). In the meta-analysis study of the link between birth weight and leukemia, Hjalgrim et al. (7) observed a nonstatistically significant dose–response-like association (OR = 1.29, 95% CI = 0.80 to 2.06 per 1-kg increase in birth weight) between birth weight and AML risk. However, results of the four AML studies included in the meta-analysis varied markedly and, accordingly, the risk estimate was difficult to interpret.

We note with interest that high birth weight has also been linked, albeit inconsistently, to the development of other childhood tumors such as Wilms tumor and neuroblastoma (13,18,49,50). This result could indicate a more general phenomenon, namely that birth weight was associated with various childhood malignancies (36).

Increasing gestational age was associated with a statistically significantly reduced risk of B-precursor ALL (OR = 0.87, 95% CI = 0.81 to 0.94) but apparently not with that of T-cell ALL (OR = 1.12, 95% CI = 0.88 to 1.41). The association of gestational age with risk of AML was not statistically significant but was of the same magnitude as the effect seen for ALL overall. Reports about the association between gestational age and leukemia risk are inconsistent (9,10,14,16,48,5154). We analyzed data from population-based registers, which gives high credibility to our results. The multivariate analyses revealed increasing birth weight and decreasing gestational age to be independently associated with increased ALL risk. Univariate analyses placed these estimates toward the null because of the strong association between birth weight and gestational age. The independent effects of birth weight and gestational age would be consistent with the hypothesis that the association between birth weight and ALL risk reflects proliferative stress; i.e., the shorter the time to a given birth weight, the greater the proliferative stress and the greater the leukemia risk.

We observed that increasing birth order position was statistically significantly associated with the risk of B-precursor ALL but found no association between birth order and T-cell ALL or AML. Reports about the association between birth order and ALL risk are inconsistent (17), but a similar association between birth order and overall ALL risk has recently been reported in a register-based case–control study of 2000 children aged 1–5 years with leukemia (55). It has been hypothesized that the characteristic peak of B-precursor ALL incidence among 1- to 5-year-old children in countries with a high socioeconomic status is caused by a delayed exposure to common infections (8). Birth order may serve as an indirect measure of the infectious burden in early childhood, with children having a low birth order position being less likely to be exposed to common infections early in life than children with a higher birth order position (56). Although our results would be consistent with a delayed infection hypothesis for development of B-precursor ALL, other factors, such as the use of day care and vaccination programs, are also likely to affect the burden of infectious diseases. Alternative explanations for the effect of birth order on ALL should therefore not be ruled out.

The quality of the data used in this study gives high credibility to our results. We linked information from the different population-based health registers in the Nordic countries by the unique person-number systems with information retrieved from the Medical Birth Registers that contains mandatorily reported and continuously updated data on births (2629). Control children were randomly selected from the entire population at risk of leukemia in each country. All patients in the Nordic countries have free and easy access to public health care, and children with leukemia are treated according to common Nordic protocols in centralized pediatric hematologic and oncologic centers (21). This system ensures the complete registration of patients in the NOPHO acute leukemia database, and follow-up of each patient with respect to treatment status also ensures a high degree of validity of all registered diagnoses. Therefore, misclassification of case patients and selection bias of case patients and control subjects are considered negligible problems in this study. Moreover, any misclassification on the exposure variables birth weight, gestational age, and birth order would be non-differential and therefore would tend to underestimate the risk of leukemia associated with high birth weight (57).

Surveys have revealed a 10-fold variation in the incidence of childhood leukemia worldwide (58). Although the geographic incidence of childhood leukemia appears to be associated with level of socioeconomic development (59), studies of the association between socioeconomic status and leukemia risk on the individual level are not consistent (55,60). We also note that the association between birth weight and ALL risk remained statistically significant in previous investigations after adjustment for socioeconomic status (10,11). Therefore, we find it unlikely that the observed association between birth weight and leukemia risk in this investigation should result from socioeconomic confounding.

Cases with Down syndrome, the genetic disorder most frequently associated with leukemia risk, were excluded from our analyses. Compared with other children, children with Down syndrome have a 30-fold increased ALL risk and more than a 100-fold increased AML risk (17). We did not have information on Down syndrome for the control subjects. Down syndrome children tend to have lower birth weights than normal children (61); however, only one birth in a thousand will result in a child with Down syndrome (62). Thus, it is unlikely that exclusion of children from the control group would have changed the observed relationship between birth weight and ALL risk.

In conclusion, we observed a statistically significant dose–response association between birth weight and ALL risk that was similar for all ALL subtypes and all diagnostic age groups. Moreover, we found that children with ALL did not weigh more at birth than their siblings but that, children with ALL from case families weighed more than those from control families. Our results are compatible with the hypothesis that high birth weight modifies the risk of ALL, either through proliferative stress and/or by increasing the number of cells at risk of leukemia-associated genetic aberrations.


    NOTES
 Top
 Notes
 Abstract
 Introduction
 Patients and Methods
 Results
 Discussion
 References
 
Supported by the Danish Cancer Society (grant DP01083), the Danish Children’s Cancer Foundation, the Dagmar Marshall Foundation, the Augustinus Foundation (grant 2-1919), the Aase and Ejnar Danielsen Foundation (grant 103810), the Ejnar Willumsens and Wife Memorial Foundation (grant 800.211), and the A. P. Moeller Foundation for the Advancement of Medical Science (grant 01075).


    REFERENCES
 Top
 Notes
 Abstract
 Introduction
 Patients and Methods
 Results
 Discussion
 References
 

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Manuscript received March 4, 2004; revised August 10, 2004; accepted August 20, 2004.


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