Birth Weight and Childhood Cancer Deaths 1,2

Transcription

1 Birth Weight and Childhood Cancer Deaths 1,2 Diane E. Eisenberg 3,4 and Tom Sorahan 5,6 ABSTRACT -For investigation of the hypothesis that elevated birth weight characterizes children dying from cancer, birth weights of over 3,000 children who died of cancer were compared with those of matched controls alive at the time of the case's death. Analysis revealed no consistent pattern of association between birth weight and subsequent cancer death. Girls who died of solid tumors between ages 1 and 10 had a significantly higher mean birth weight and an excess of high birth weights separate from the influence of maternal age, birth order, and socioeconomic status. Boys who died of solid tumors between birth and 2 years had a significantly lower mean birth weight and exhibited a significant, progressive linear decrease in representation across the birth-weight distribution in the controlled analysis. However, this study, the largest to date, did not support previous claims that birth weight may be an important predictor of childhood cancer risk.-jnci 1987; 78: Numerous investigations have found elevated birth weight among children with leukemia and other cancers (1-7). However, only one of these studies (5) used an appropriate comparison group (i.e., one that survived past infancy and thereby excluded the low-birth-weight neonatal deaths) and also controlled for potentially confounding effects of maternal age, birth order, and socioeconomic status. The present report, the largest study to date, analyzes the relation between birth weight and childhood cancer before age 15 in Great Britain from 1965 to Matched controls, alive at the time of the case child's death, constituted an appropriate comparison group, and information on confounding factors was available for both cases and controls. To investigate the hypothesis that elevated birth weight characterizes children dying from cancer, we compared mean birth weights of case and control children and also employed a Mantel Haenszel approach to contrast the birth-weight distributions of these groups. The large study population permitted detailed comparisons among cancer diagnostic categories and subgroups defined by the child's age at death. SUBJECTS AND METHODS The data were derived from an ongoing collaborative study, the Oxford Survey of Childhood Cancers, which has been described previously (8). From 1953 to the present, the Oxford Survey has obtained information on children who have died from cancer in England, Scotland, and Wales from death registrations, hospital or general practitioner records, and interviews with parents. Information was similarly obtained on control children who were alive at the time of the case child's death. The control was selected from a list of children registered as being born in the same locality as the case, of the same sex, in the same month or half-year. An interview was arranged with one control mother from this list and was conducted by the same survey doctor who interviewed the case mother. For a portion of the matched pairs in which the case died between 1965 and 1970, birth weights had been coded on computer records: From a total of 5,747 singleborn case-control pairs of that period, birth weights were available for 67% (3,868) of the cases and 64% (3,691) of the controls. The analyses described below used subsets of these groups with adequate covariable information. The birth weights, obtained from the prenatal clinic, the general practitioner, or the mother, were recorded in ounces. Two approaches were used to bring to light any relation between birth weight and childhood cancer mortality. First, a series of paired t-tests (9) was performed to identify any significant differences in the birth weights of matched case-control pairs. There were 2,898 matched pairs for which birth weights were available for both members of the pair. For each sex, separate tests were performed for leukemia cases and controls (1,207 pairs), other RES neoplasms (264 pairs), and solid tumors (1,424 pairs). These analyses were done on all cancer deaths occurring before age 15 and then on subgroups defined by the cancer case's age at death. Three deathage subgroups were defined: deaths in the 1st year of life, deaths between 1 and 10 years of age, and "peak" death ages, which were defined specific to this study population as the 3-year age span in which one-fourth to one-third of the deaths occurred, i,e., for leukemia, ABBREVIA non USED: RES = reticuloendothelial system. I Received May 8, 1986; revised November 7, 1986; accepted December 8, Supported by the Dissertation Travel Award, University of Wisconsin-Madison Graduate School; and by the Department of Social Medicine, University of Birmingham, England. 3 Institute for Environmental Studies and Department of Preventive Medicine, University of Wisconsin-Madison, Madison, WI Address reprint requests to Ms. Eisenberg at Room 253 Hygiene Laboratory, University of Wisconsin-Madison, Madison, WI Cancer Epidemiology Research Unit, Department of Social Medicine, University of Birmingham, Edgbaston, Birmingham BI5 2TT England. 6 We thank Dr. Alice Stewart for initiating and sustaining the Oxford Survey of Childhood Cancers that provided the data we analyzed, Mr. George Kneale for comment and illumination regarding the statistical methods, Professor George Knox and Dr. Lorraine Meisner for interest and encouragement, Mr. Robert Lancashire for patient programming assistance, and Ms. Anne Walker and Ms. Pamela Goedel for assistance in preparing the manuscript JNCI, VOL. 78, NO.6, JUNE 1987

2 1096 Eisenberg and Sorahan ages 3-5; for other RES neoplasms, ages 6-8; and for solid tumors, birth-2. Since the paired t-test approach could only discover a difference in the mean birth weight of cases and controls, a more thorough analysis was devised to point out any contrasts between the birth-weight distributions of case and control children. The continuous birth-weight distribution was divided into six strata (see tables 3 and 4), and a series of t-statistics was calculated to compare observed and expected frequencies of children in each stratum with the use of Mantel-Haenszel cross-classification techniques (10) to control for the potentially confounding factors, maternal age, birth order, and socioeconomic status. It was decided to control for these factors inasmuch as they influence birth weight and had been shown to influence childhood cancer risk in a number of previous studies (1,4,8). Mantel-Haenszel estimates of odds ratios were also calculated for each stratum, and these results are shown in the tables that follow. The third birth-weight stratum was chosen as the referent category, with odds ratio set at 1.0, since the large numbers in this intermediate group would provide stable estimates of odds ratios in the remaining categories. Two overall contrasts of the birth-weight distributions of cases and controls were done: an asymptotic normally distributed test statistic that measures any tendency of the risk to increase or decrease progressively across birth weight strata (11) and a chi-square test for homogeneity of risk across strata. Individual matching was ignored; expected frequencies were calculated from the combined distribution of cases and controls, with the use of all controls with a complete set of covariables. In this manner, the birthweight distribution of 3,070 controls was contrasted with that of cases with covariable information, i.e., the 1,304 leukemias, 286 other RES neoplasms, and 1,537 solid-tumor deaths that occurred before age 15. As in the matched-pair analysis, these analyses were performed separately by sex and then by subgroups defined by the case's age at death. In the examination of death-age subgroups, only those controls whose matched case had died at the age of interest were included in the calculation of the expected figures. Table 1 characterizes the study population by cancer type and variables controlled for in the Mantel-Haenszel stratified analysis. RESULTS None of the analyses revealed a systematic pattern of high or low birth weight among childhood cancer cases. The matched-pair analysis found no consistent direction in the difference between birth weights of matched case-control pairs. Out of 24 t-tests shown in table 2, 3 subgroups revealed differences in birth weight that were significant at the 5% level of probability. For solid tumors, girls who died between ages 1 and 10 were slightly heavier at birth than controls, whereas boys who died at the peak ages 0-2 were lighter. For leukemia there was no significant difference between mean birth TABLE I.-Characteristics of the study population and covariate stratification used in the Mantel-Haenszel analysis G Cancer Diagnosis of Cases No. Leukemia (ICD b No ) 1,304 Other RES neoplasms (ICD No ; ) 286 Solid tumors (ICD No ; ) 1,537 3,127 Controls Birth order Cases 1, , ~ ,070 3,127 Controls Maternal age Cases 38 -S , : ,070 3,127 Controls Social class c Cases 107 I II 589 1,838 III 1, IV V 167 3,070 3,127 G For the purposes of stratification in the Mantel-Haenszel analysis, birth orders 3 and 4 were combined and social classes 1 and 2 were combined. b ICD = International Classification of Diseases, Eighth Revision. C This stratification follows the Registrar General's classification of social class based on the father's occupation. weight of cases and controls. For RES neoplasms other than leukemia, boys who died between ages 6 and 8 were lighter at birth. Table 2 shows the birth-weight differences to be of small magnitude and with no apparent pattern in subgroups defined by cancer type, gender, and age at death. The stratified analysis controlling for birth order, maternal age, and socioeconomic status revealed no distinctive birth-weight distribution for subsequent cancer cases. Table 3 shows two significant contrasts between observed and expected numbers in birth-weight strata out of 36 comparisons among all cancer deaths from birth to age 15; boys who died of solid tumors were underrepresented (t = -2.4) in the fourth stratum ( oz) and overrepresented (t = 2.0) in the fifth stratum ( oz). The lack of significance in any of the across-strata tests-for trend or homogeneity-indicates that cases had no distinctive overall pattern of birthweight distribution. Table 4 sets out results of the stratified analyses of subgroups defined by the case's age at death. No striking contrast between case and control birth weights is evident. Out of 96 comparisons, 3 are significant at the JNCI. VOL. 78, NO.6, JUNE 1987

3 Birth Weight and Childhood Cancer 1097 TABLE 2.-Paired t-tests of birth-weight differences of matched cases and controls by cancer type, death age, and sex Mean difference Cancer type, category, and subgroup a Degrees of freedom, n= 1 t-test in ounces, case wt-control wt Leukemias Females (peak) Males (peak) Solid tumors Females b ' 0-2 (peak) Males (peak) h d Other RES neoplasms Females (peak) Males (peak) b -8.5 e a Peak death age was defined specific to this study population as the 3-yr age span in which one-fourth to one-third of the deaths occurred. b P<.05. '95% confidence interval (0.7, 5.4). d 95% confidence interval (-7.2, -0.3). e 95% confidence interval (-15.8, -1.2). 5% level of probability: for leukemia, girls who died DISCUSSION between ages 3 and 5 are overrepresented (t = 2.3) in the fourth stratum (1l5-127 ( oz) and underrepresented (t = 2.1) in the sixth stratum (2:: 140 oz); for solid tumors, girls who died between ages 1 and 10 are overrepresented Genetics textbooks cite birth weight as the most clearcut example of a human character subject to stabilizing selection (12) in that neonatal mortality exhibits a para (t = 1.9) in the sixth stratum (2::140 oz). There is a sigbolic curve with respect to birth weight, i.e., babies that nificant (t = -2.2) linear decrease across birth-weight weigh between 5.9 and 10 lb are more likely to survive strata for boys who died from solid tumors from birth to the first 4 weeks of life than those whose birth weight 2 years. lies outside these limits. Yet birth weight's influence on These scattered instances of statistical significance seem to reflect sample size in large part. A scan across strata in tables 3 and 4 suggests no characteristic birthweight pattern for cases stratified by cancer type or age at death. survival fades quickly in infancy. Gibson and McKeown (13) found that infant mortality did not vary with birth weight after the 1st month of life. In this context the observed association between high birth weight and childhood cancer has been interpreted to be indirect via JNCI, VOL. 78, NO.6, JUNE 1987

4 1098 Eisenberg and Sorahan TABLE 3.-0dds ratios a for childhood cancer death, ages 0-15, across the birth-weight distribution Cancer type Birth weight, oz Trend Chi-square and category No. ~ ~140 statistic b homogeneity Leukemia Girls c Boys Other RES neoplasms Girls Boys Solid tumors Girls Boys d 1.2 d a These odds ratios are calculated by a Mantel-Haenszel stratified analysis, controlling for birth order, maternal age, and socioeconomic status. The birth-weight distribution of 3,070 controls matched to all cancer cases served as the comparison. b Asymptotic normally distributed. c The third birth-weight stratum was chosen as the referent category, with odds ratio set at 1.0, since the large numbers in t1~is category ensure stable odds ratio estimates for the remaining categories. d P<.05. some prenatal influence that predisposes to both outcomes, e.g., maternal diabetes or in utero x-ray exposure (5, 7). An association between high birth weight and childhood cancer was first noted more than 20 years ago (1), and subsequent studies have repeated and refined the observation (2-7). The present study, the largest to date, gives little support for such an association, although TABLE 4.-0dds ratios a for childhood cancer death, by age at cancer death, across the birth-weight distribution Cancer type, subgroup, Birth weight, oz Trend No. Chi-square and category statistic b homogeneity ~ " ~140 Leukemia Infant deaths d Girls Deaths yr Boys Girls Boys Peak deaths, 3-5 yr e Girls Boys Other RES neoplasms Deaths yr Girls Boys Peak deaths, 6-8 yr Girls Boys Solid tumors Infant deaths Girls Boys Deaths yr Girls Boys Peak deaths, 0-2 yr Girls Boys a We calculated these odds ratios by a Mantel-Haenszel stratified analysis controlling for birth order, maternal age, and socioeconomic status. The birth-weight distribution of controls matched to all cancer cases who died at the age of interest served as the comparison. b Asymptotic normally distributed. C The third birth-weight stratum was chosen as the referent category, with odds ratio set at 1.0, since the large numbers in this category ensure stable odds ratio estimates for the remaining categories. d Due to small numbers, this infant death category includes cases who died from leukemia or other RES neoplasms. e Peak death age was defined specific to this study population as the 3-yr age span in which one-fourth to one-third of the deaths occurred. Ip<.05. lnci, VOL. 78, NO.6, JUNE 1987

5 Birth Weight and Childhood Cancer 1099 there are some significant findings among specific subgroups. Our study supports and counters previous reports on several points. The first report, by MacMahon and Newill (1), compared birth weights of 2,653 children who had died of cancer from 1947 to 1958 at ages 0-11 years. In each diagnostic category examined, children who died of cancer had a higher mean birth weight and a larger proportion of high birth weights (>8.5 Ib) than the comparison series. Our study does not confirm either observation. The authors themselves considered the higher birth weights of their cancer series to be an artifact: Their comparison series included low-birth-weight neonatal deaths, whereas the cases, having survived long enough to exhibit cancer, excluded these. Subsequent studies focused on children with leukemia. Iversen (2) reported a deficit of low birth weight among 258 children with acute leukemia, ages 0-14, compared to the national average for all live births. Our much larger study population exhibited the opposite relation (see tables 2 and 3). Jackson et al. (3) found that in 29 out of 42 twins discordant for leukemia, ages 0-14, the twin who died from leukemia weighed more at birth. Stratifying by sex composition and birth order of the twin pair, they found this difference to be significant and confined to the female leukemic twins. Stratifying by age at death, they found that the leukemic twin was more often heavier than the cotwin at birth for ages at death 0-14, 0-4, and 5-14, and that this difference was significant in the first two strata. Wertelecki and Mantel (4) found that children treated for leukemia consistently weighed more than the mean birth weight of their siblings and had significantly higher birth-weight ranks within their sibships than expected. Our study does not contrast birth weights of twins or siblings; our leukemia cases show no evidence of higher birth weight than the controls. Fasal et al. (5), in the largest prior study of this hypothesis, compared birth weights of 802 children who had died from leukemia between ages 1 and 9 with birth weights of children who survived infancy. Controlling for maternal age and social class, they found that girls of high birth weight (>8.5 Ib) had a significant doubled risk of leukemia mortality. This relation was not observed among the boys, and there was no progressive increase in risk from lower to higher birth weights. Our female leukemia case series lends no support to this observation; our female solid-tumor cases do exhibit a significant excess of high birth weight (see table 4). Recent studies have used tumor registries to examine birth weights of incident childhood cancer cases. Hirayama (6) noted a small, significant excess risk of leukemia, testicular tumors, teratoma, and Wilms' tumor among high-birth-weight children whose cancers were diagnosed between birth and 2 years. There was no birthweight association for children with cancers diagnosed after age 2. Following up on this study, Daling and coauthors (7) examined the birth-weight distribution of children with cancer by age at diagnosis. Controlling for sex and year of birth, they contrasted the incident cases' representation across five birth-weight strata with that of all liveborn children in Washington State. They found that of 178 children with cancers diagnosed before age 2, 23% (n = 40) weighed over 4,000 g compared to 13% of the control series; the birth-weight distribution of these early cancers differed significantly from the controls. Although boys had an excess of high birth weights, the excess reached statistical significance only among girls. This relationship was not observed in boys or girls with tumors diagnosed after age 2. The early cancers associated with high birth weights were Wilms' tumor, leukemia, and neuroblastoma. Our study benefits from several strengths compared to this most recent study. Our case series exceeds 3,000; while this figure includes cancers up to age 15, the number of cancers diagnosed before age 2 is more than double the corresponding number in the study of Daling et al. Our analysis controls for birth order, maternal age, and socioeconomic status. Daling cites these as possible explanatory factors for their observed association. Our control series has survived infancy and thus excludes the low-birth-weight neonatal and infant deaths, which might create an artifactual relation between an excess proportion of high birth weight and childhood cancer. Their examination of incident cases according to age at diagnosis versus our examination of cancer deaths according to age at death makes the studies somewhat disparate. Yet this difference does not explain our discordant results. For example, although the 5-year survival for acute lymphoid leukemia in childhood has improved considerably in recent decades, it was still below one-third in in northwest England (14). Had there been an association between high birth weight and early incident leukemia cases in our study population, as in Daling's, our large mortality analysis would have discovered it in later death ages. On the contrary, we found a significantly low mean birth weight, and a significant decreasing trend across birth-weight strata, among boys who died of solid tumors before age 2. The only excess of high birth weight we observed occurred among girls who died of solid tumors between ages 1 and 10. Our study is handicapped in that it examines deaths rather than incident cases. Case survival rates changed markedly over the decades in which our study and the others were conducted (14), making contrasts between the studies less interpretable. Our analysis is also weakened by its reliance on only two-thirds of the casecontrol series for , i.e., those for which birth weights were computer coded. In summation, studies of the birth weight-childhood cancer relation lack comparability. Some studies did not control for confounding factors; others were based on small numbers. The speculative hypotheses put forward to explain this elusive, variable relation include a) a common prenatal factor (e.g., maternal diabetes, in utero x-ray, or immunologic mechanisms) that predisposes to both high birth weight and childhood cancer (4, 5, 7) and b) that high birth weight itself promotes the development of incipient childhood cancers (4). JNCI, VOL. 78, NO.6, JUNE 1987

6 1100 Eisenberg and Sorahan Neither hypothesis suggests why the relation would be restricted to a particular subgroup defined by sex, cancer type, and age at diagnosis. Also, although our study does not control for obstetric radiation, this omission would only tend to produce a spurious birth weightcancer relation rather than the null effect in fact found. Our study does not confirm previous observations but only identifies new nonsystematic relations between birth weight and subsequent cancer risk. We conclude that birth weight is not an important indicator of a child's cancer risk. REFERENCES (1) MACMAHON B, NEWILL VA. Birth characteristics of children dying of malignant neoplasms. J Nat! Cancer Inst 1962; 28: (2) IVERSEN T. Leukaemia in infancy and childhood. A material of 570 Danish cases. Acta Paediatr Scand [Suppl] 1966; 167: (3) JACKSON EW, NORRIS FD, KLAUBER MR. Childhood leukemia in California twins. Cancer 1969; 23: (4) WERTELECKI W, MANTEL N. Increased birth weight in leukemia. Pediatr Res 1973; 7: (5) FASAL E, JACKSON EW, KLAUBER MR. Birth characteristics and leukemia in childhood. J Natl Cancer Inst 1971; 47: (6) HIRAYAMA T. Descriptive and analytical epidemiology of childhood malignancy in Japan. In: Kobayashi N, ed. Recent advances in managements of children with cancer. Tokyo: The Children's Cancer Association of Japan, 1980: (7) DALING JR, STARZYK P, OLSHAN AF, et al Birth weight and the incidence of childhood cancer. JNCI 1984; 72: (8) BITHELL JR, STEWART AM. Pre-natal irradiation and childhood malignancy. Br J Cancer 1975; 31: (9) SNEDECOR GW, COCHRAN WG. Statistical methods. Ames, IA: Iowa State Univ Press, (10) MANTEL N, HAENSZEL W. Statistical aspects of the analysis of data from retrospective studies of disease. J Nat! Cancer Inst 1959; 22: (1I) MANTEL N. Chi-square tests with one degree of freedom: Extensions of the Mantel-Haenszel procedure. J Am Stat Assoc 1963; 58: (12) CAVALLI-SFORZA L, BODMER WF. The genetics of human populations. San Francisco: Freeman, (13) GIBSON JR, McKEOWN T. Observations on all births (23,970) in Birmingham, III. SurvivaL Br J Soc Med 1951; 5: (14) BIRCH JM, SWINDELL R, MARSDEN HB, et al Childhood leukaemia in North West England : Epidemiology, incidence and survival Br J Cancer 1981; InCl 43: JNCI, VOL. 78, NO.6, JUNE 1987