Neonatal hypothermia is associated with an increased mortality risk for 28 days. There are few community-based data on specific risk factors for neonatal hypothermia. Estimates of association between neonatal hypothermia in the community and risk factors are needed to guide the design of interventions to reduce exposure.A cohort of 23,240 babies in rural southern Nepal was visited at home by field workers who measured axillary temperatures for 28 days . The cumulative incidence of hypothermia  was examined for any association with infant characteristics, care practices and parental, household, socioeconomic and demographic factors. Estimates were adjusted for age and ambient temperature.n = 2342) were observed with temperatures of < 35.0°C. Adjusted prevalence ratios (Adj PR) were increased among those who weighed < 2000 g [Adj PR = 4.32 ] or < 1500 g [Adj PR = 11.63 ] compared to those of normal weight (> 2500 g). Risk varied inversely along the entire weight spectrum: for every 100 g decrement hypothermia risk increased by 7.4%, 13.5% and 31.3%% for babies between 3000 g and 2500 g, 2500 g and 2000 g and < 2000 g, respectively. Preterm babies (< 34 weeks), females, those who had been first breastfed after 24 h and those with hypothermic mothers were at an increased risk. In the hot season the risk disparity between smaller and larger babies increased. Hypothermia was not associated with delayed bathing, hat wearing, room warming or skin-to-skin contact: they may have been practiced reactively and thereby obscured any potential benefit.Ten percent of the babies (In addition to season in which the babies were born, weight is an important risk factor for hypothermia. Smaller babies are at higher relative risk of hypothermia during the warm period and do not receive the protective seasonal benefit apparent among larger babies. The need for year-round thermal care, early breastfeeding and maternal thermal care should be emphasized. Further work is needed to quantify the benefits of other simple neonatal thermal care practices. Neonatal hypothermia is widely recognized as an important contributing factor to neonatal morbidity, especially in low and middle income countries ,2 and haFew population-based estimates are available from settings where the majority of babies are born at home, where the risk for hypothermia is presumably higher than in hospital ,14-16. IContext-specific information on the risk factors for neonatal hypothermia is necessary in order to ensure that community-based neonatal health promotion packages include thermal care interventions for the newborn. Community-based trials of chlorhexidine cleansing interventions in southern Nepal ,20 incluData for this analysis were collected during large, community-based, placebo-controlled, randomized trials of the efficacy of chlorhexidine interventions  on neonatal mortality; the details of those trials have been previously published ,20. BetwMahdeshi, originating from the plains or Pahadi, originating from the hills), material and/or livestock ownership and parental characteristics - such as maternal birth history, parity, parental literacy, education and occupation - collected either during a population census at baseline or during the single antenatal enrollment visit. In addition, on two follow-up visits (days 1 and 14) field workers collected information on various potential determinants of hypothermia related to neonatal care practices during labour and delivery, immediately after birth and during the postnatal period. These included: birth location; type of birth assistant ; timing of placental delivery; sex and weight of the infant ; gestational age ; breastfeeding initiation time , as were those born to mothers assisted at birth by non-skilled personnel [PR = 1.47 ] and babies of mothers reporting at a > 30 min interval between the birth and delivery of the placenta. Upper caste, ethnic group and variables indicating improved socioeconomic status provided a greater protection from hypothermia.Hypothermia was almost 50% more prevalent among babies for whom breastfeeding was delayed beyond 24 h [PR = 1.49 ) and 75% more prevalent among babies for whom the cord was not cut until after the placenta was delivered [PR = 1.74 ]. Risk was marginally higher among babies bathed after birth, an almost universal practice in this setting  and < 2000 g [31.3% ].Given the strength of the association with infant weight, the risk associated with incremental (50 g) changes in weight was examined Figure . After mP-values: < 0.001, 0.06 and 0.003, respectively). In the warm season, relative to newborns ≥ 2500 g, the adjusted prevalence of hypothermia was 23.46  and 5.05  times greater in the lowest (< 1500 g) and second-lowest (2000 g-2499 g) weight categories, respectively. The parallel ratios of adjusted prevalence (Adj PR) during the cold season were less: 4.04  and 3.80 , respectively. Similarly, wearing a hat was protective only during the cold season [Adj PR = 0.71 (0.52 - 0.96)].The season during which a child was born was examined for interaction terms for birth weight category, delivery location, birth attendance with skilled personnel, hat-wearing during the first 14 days and caste and ethnicity. There was evidence that the season during which babes were born modified the association between hypothermia and the two lowest birth weight categories and hat wearing by the infant , the weight of the infant was the most important independent factor related to neonatal hypothermia. Almost 30% of infants had low birth weight < 2500 g) and those between 2000 g -2499 g and 1500 g -1999 g were at a 1.49 and 4.32 times greater risk of hypothermia, respectively, after adjusting for a multitude of factors. VLBW babies were 11.63 times more likely than normal weight infants to have axillary temperatures < 35.0°C, although the proportion of these babies was very low . When examined as a continuous variable, 100 g decrements in weight were associated with substantial increases in risk of hypothermia along the entire spectrum of underweight babies and continued up to 3000 g. Such associations have been documented previously among hospital-born newborns ,25,26.00 g and All newborn babies are prone to heat loss, particularly in the first minutes and hours after birth ,2. PreteThese weight measures (precise to +/- 2 g) are more objective than gestational age, which was estimated from the date of the last menstrual period reported by mothers at enrollment at approximately 6 months. Thus, there is more scope for misclassification of gestational age, which could result in estimates of the association with hypothermia that are biased downwards . While wAdj PRs for the lowest birth weight category were five times higher in the warm season than in the cold. This indicates that, while the overall risk of hypothermia is lower during this time relative to the cold months,  low birtdiffering intent between boys and girls (as being preventive for boys but therapeutic for girls), these sex-related associations could arise even after adjustment.The strong association between hypothermia and the sex of the newborn is not easily explained. There may be unmeasured practices that differ according to the sex of the newborn which may change the risk profile . AlternaDelayed breastfeeding was associated with hypothermia after adjustment: early breastfeeding may directly reduce risk through this additional direct contact. In addition, early breastfeeding has been associated with reduced neonatal mortality ,31 potenWhile kangaroo-mother-care or skin-to-skin contact is recommended as part of a package of targeted care for low birth weight or preterm infants in facilities  debate cIf adequate thermal protection is not begun immediately after birth, the newborn will lose heat rapidly, with the skin temperature losing up to 4°C in the first minutes . Thus, iThese data provide an in-depth examination of the risk factors associated with neonatal hypothermia as measured by axillary temperatures collected at home from 23,240 infants. The data highlight the importance of thermal care interventions, especially those targeted towards low birth weight infants. Focused behaviour change messages and skill building on the basic components of thermal care, including early breastfeeding initiation, skin-to-skin contact for all infants and delayed bathing have been promoted in Shivgarh, India and have been resulted in a 54% reduction in neonatal mortality . The relAdj PR: adjusted PR; CI: confidence interval; LBW: low birth weight; PR: prevalence ratio; VLBW: very low birth weight; WHO: World Health Organization.The authors declare that they have no competing interests.LCM contributed to the concept, design, conduct and analysis of the study and also wrote the manuscript. JK contributed to the overall study concept, design, interpretation and provided a critical review of the manuscript. SKK and SCL contributed to the design and implementation of the study, provided an oversight of data collection and critically reviewed the manuscript. GLD contributed to the overall study concept, interpretation and critical review of the manuscript. JMT provided inputs on the overall concept and design of the study, the interpretation of the results and a critical review of the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1741-7015/8/43/prepub


Genetic and environmental factors have important roles in multiple sclerosis (MS) susceptibility. A clear parent of origin effect has been shown in several populations, perhaps resulting from factors operating during gestation. Preterm birth  has been shown to result in long-term health problems, including impaired neurological development. Here, in a population-based cohort, we investigate whether preterm birth increases the risk to subsequently develop MS.We identified 6585 MS index cases and 2509 spousal controls with preterm birth information from the Canadian Collaborative Project on Genetic Susceptibility to MS. Rates of individuals born preterm were compared for index cases and controls.p = 0.41.There were no significant differences between cases and controls with respect to preterm births. 370 (5.6%) MS index cases and 130 (5.2%) spousal controls were born preterm, Preterm birth does not appear to contribute to MS aetiology. Other factors involved in foetal and early development need to be explored to elucidate the mechanism of the increased risk conferred by the apparent maternal effect. Multiple sclerosis (MS) is a chronic inflammatory disease of the central nervous system (CNS) characterized by myelin loss, varying degrees of axonal pathology and progressive neurological dysfunction . With a A parent-of-origin effect  has been repeatedly observed in MS, based on data from studies of half-siblings , sibshipA "preterm" birth is defined as occurring at less than 37 weeks gestational age . PretermIn MS, plaque load does not correlate with axonal loss , the accHere, in a population-based cohort, we investigate whether preterm birth increases the risk to subsequently develop MS. To the best of our knowledge, this is the first study on this specific topic.Cases and controls were identified through the population-based, longitudinal Canadian Collaborative Project on Genetic Susceptibility to Multiple Sclerosis (CCPGSMS), the methodology of which has been previously described ,17. BrieSpecific to this study, preterm data was collected by telephone interview with mothers of MS index cases and spouse controls . Each CCPGSMS index case and spouse control (if available) was asked if he/she had a biological mother who could provide information on the individual's early life. If a positive answer was received the biological mother of the case/control was contacted by a CCPGSMS site research nurse and administered a standardised questionnaire, details of which are described in . CCPGSMSThe Chi squared test was used to assess significance when comparing index cases and controls for the rate of preterm births. The Student's t-test was used to look for differences in average gestational age. Logistic regression  was used to assess the effect of preterm birth, gender and the interaction thereof.Complete information from maternal informants on weeks gestation at birth as well as birth weight, duration of hospital stay, and the use of an incubator was available for 6585 index cases and 2509 spousal controls. The clinical and demographic details of the index cases and controls are shown in Table p = 0.41 (see Table p = 0.67 and p = 0.31 respectively) in the rates of preterm birth were observed when comparing female cases  and controls  – or male cases  and controls  (Table p < 1 × 10-16) but no effect of being born preterm (p = 0.65), or an interaction effect between sex and preterm birth (p = 0.31) on the likelihood of developing the disease.Three hundred and seventy (5.6%) MS index cases and 130 5.2%) spousal controls were born preterm. These values were not significantly different; % spousalp = 0.25) – see Table p = 0.82; average gestational age male index cases born preterm = 34.2 weeks, average gestational age male controls born preterm = 33.9 weeks, p = 0.52).The average gestational age of preterm MS index cases was 34.6 weeks compared to 34.2 weeks for controls (data not shown).There were no differences in the number of individuals born preterm between MS index cases and controls when stratifying by month of birth or when stratifying MS index cases by clinical course  effect shown by several studies ,5,6. IntHere, we observed no significant differences in preterm births for cases or controls, even when stratifying by sex. The mean gestational age for preterm cases and controls was also similar Table . Furtherdata not shown). Greater understanding of foetal development and early maternal influences on the foetus and child is needed before the early origins of multiple sclerosis can be assessed to a greater extent.The clinical definition of preterm birth may not take into account detrimental aspects of foetal development associated with premature birth measured in days rather than weeks that may contribute to the onset of MS in adulthood. Development of gyri and sulci has been shown to take place well into 40 weeks of gestational age , and henThis study does not rule out maternal risk factors associated with preterm birth, for example smoking and infection, from MS disease pathogenesis, as these factors may still have a role independent of any effect they may have on premature birth.Our study may be limited in that maternal recall of data may not be as accurate as data from clinical records etc. Additionally, as we erred on the side of caution when classifying individuals as preterm, we may have missed those who were born borderline preterm and hence we may be underestimating the number of preterm births in our cohort. However, this will apply equally to cases and controls and maternal recall of preterm birth has been shown to be accurate enough for epidemiological use .It should be remembered that parent of origin effects can arise in a number of ways, and likely also by other means as yet undiscovered. These mechanisms include genomic imprinting  and micrCCPGSMS: Canadian Collaborative Study Group on the Genetic Susceptibility to Multiple Sclerosis, MS- Multiple SclerosisThe authors declare that they have no competing interests.GCE and ADS conceived and designed the experiments. SVR, WV, DAD, GCD, SMO, IMY, MC, GCE and ADS analyzed the data and wrote the paper. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


In rural Gambia, birth season predicts infection-related adult mortality, providing evidence that seasonal factors in early life may programme immune development. This study tested whether lymphocyte subpopulations assessed by automated full blood count and flow cytometry in cord blood and at 8, 16 and 52 weeks in rural Gambian infants (N = 138) are affected by birth season , birth size or micronutrient status.Geometric mean cord and postnatal counts were higher in births occurring in the WET season with both season of birth and season of sampling effects. Absolute CD3+, CD8+, and CD56+ counts, were higher in WET season births, but absolute CD4+ counts were unaffected and percentage CD4+ counts were therefore lower. CD19+ counts showed no association with birth season but were associated with concurrent plasma zinc status. There were no other associations between subpopulation counts and micronutrient or anthropometric status.in utero) and subsequent samples, and is not explained by nutritional status. These findings are consistent with the hypothesis than an early environmental exposure can programme human immune development.These results demonstrate a seasonal influence on cell counts with a disproportionate effect on CD8+ and CD56+ relative to CD4+ cells. This seasonal difference was seen in cord blood (indicating an effect  Using demographic data from West Kiang in rural Gambia, collected since 1949, we have previously demonstrated a profound birth-season bias in adult deaths, with a large excess of early adult deaths amongst individuals born in July-December . The preThe thymus is a potential programming target that is central to the development of adaptive immunity, contributing to long-term maintenance of T-lymphocyte populations . In animThis study was designed to test the hypothesis that absolute and percentage lymphocyte subpopulation counts in this community are affected by birth-season with a greater effect on T-cell subpopulations and a maximal discrepancy between January-June (DRY) and July-December (WET) births. Here we describe changes in leucocyte, lymphocyte and CD3+, CD4+, CD8+, CD19+, CD56+ subpopulation counts in 138 rural Gambian infants.There were three deaths during follow-up: one unexpectedly at home aged 7 weeks, cause unknown, a second from malaria complicated by severe anaemia aged 34 weeks, and a third from dysentery with septicaemia aged 36 weeks. Two infants left the study during follow-up. Overall, lymphocyte subpopulation data were obtained from 453 (83%) of 545 possible samples.Mean (range) birth-weight and gestation were 2855 (2020–3900) grams and 38.6 (35.4–41.2) weeks. Birth-weight was similar in the two seasons , and this remained the case after adjusting for maternal weight. All infants were breast fed from birth, reported introduction of complementary feeds ranging from 8 to 32 weeks. Weight-for-age improved from a median z-score of -0.8 at birth to 0 at 8 weeks, but deteriorated progressively thereafter to -1.9 by 11 months. Seven blood films were positive for malaria parasites out of all the 8, 16 and 52 week visits when the infants were venesected and all samples examined for malaria parasites. There were an additional 28 positive blood films from the remaining four-weekly visits , when capillary blood was examined only in febrile infants.There was a strong effect of birth season on the total lymphocyte and leucocyte counts at all ages, including at birth . Subpopulation counts at each age are shown in Table Cord blood percentage CD4+ and CD4+/CD8+ ratios at birth were predictive of levels at 8 weeks (R = 0.3 (p = 0.003), and R = 0.28 (p = 0.0054) respectively). These, in turn and all other subpopulation percentages predicted levels at 16 weeks p ≤ 0.0001 for CD4+, CD8+, CD19+ and CD56+; p = 0.004 for CD3+). The strength of correlation between counts at 8 and 16 weeks ranged from R = 0.26 (CD3+) to R = 0.64 (CD4+).For percentage CD4+ and CD56+ counts and the CD4+/CD8+ ratio 16 week counts predicted 52 week counts (R = 0.05 (p ≤ 0.0001) for CD4+ and CD56+; R = 0.30 (p = 0.0011) for CD4+/CD8+). The CD4+ and CD19+ 8 week percentage counts at 52 weeks correlated weakly with their respective counts at 8 weeks .We found no association with current weight or weight-for-age standard deviation score for any of the subpopulations at birth, 8, 16 or 52 weeks or overall, whether expressed as absolute or percentage counts. Overall absolute CD19+ counts were positively associated with concurrent plasma zinc concentration (p = 0.0068). There were no other observed associations between any lymphocyte subpopulation and concurrent plasma zinc, retinol or vitamin C concentration.If birth weight was included in the regression model, absolute CD8+ and CD3+ counts at 8 weeks and absolute CD4+ counts at 52 weeks were positively associated with birth mid-upper-arm circumference (MUAC) . Overall absolute counts of these subpopulations were also associated with MUAC at birth . A small number of other associations were found with both head circumference and crown-heel length.Overall CD4+ and CD3+ counts were negatively associated with gestational age at birth . There was no association of the measured populations and subpopulations with gender or length of gestation at any age or overall.3 in relation to season of birth, or 2,300 leucocytes per mm3 in relation to season of sampling. These changes were accompanied by a corresponding effect on absolute lymphocyte subpopulations counts; this was selective in that it was not seen in the CD4+ subpopulation. The lack of an observed seasonal effect on percentage lymphocyte counts suggests that seasonal factors lead mainly to a non-specific increase in circulating leucocyte counts. Correspondingly, absolute but not percentage CD3+, CD8+, and CD56+ counts were positively associated with WET season birth. Seasonal fluctuations in the burden of infection provide a plausible explanation for the seasonality, since malaria transmission is intensely seasonal in The Gambia [A strong seasonal effect on circulating leucocyte counts was found in cord blood and at every age of measurement in infancy. This was manifested in consistently higher leucocyte counts in the WET season, with a magnitude of 1,600 leucocytes per mme Gambia , and male Gambia . In the e Gambia , and theThe observed selective effect on CD8+ and CD56+ cells relative to the CD4+ subpopulation could be explained by seasonal variation in exposure to viral pathogens . Studieset al found seasonal influences that were strongest in children aged 3 to 5 years, although children aged 0 to 35 months had slightly higher leucocyte counts in the wet season [et al reported lower lymphocyte counts in wet season samples. The authors found no seasonal effect on CD4+ cell counts in the younger age group, but they did report an increase in CD8+ counts relative to CD4+ counts in the wet season.Seasonal fluctuations in leucocyte and lymphocyte subpopulation counts have been reported previously in West African children . In theit season . HoweverOur results do not permit a reliable distinction of prenatal from postnatal seasonal influences, but it is notable that the effect on total lymphocyte and leucocyte counts was no less pronounced in cord blood. This confirms that a seasonal influence, infective or otherwise, operates in prenatal life. Since mothers in this community are subject to the same seasonality of infectious and parasitic disease exposure, it seems probable that the presence of maternal infection boosts leucocyte and lymphocyte counts in the fetal circulation. It is also likely that the same factors operating at the time of postnatal measurements contributed to the observed seasonality.We found no effect of birth weight on total leucocyte count, lymphocyte or subpopulation counts. However, the results showed a consistent positive influence of birth MUAC on the overall absolute CD3+, CD4+ and CD8+ counts, after adjusting for birth-weight. For CD3+ and CD8+ this result was also significant in the separate analysis of subpopulation counts at 52 weeks, and for CD4+ counts the effect was also significant at 8 weeks. These results appear robust, since they are consistent in relation to each of the measured T-cell surface markers, and significant in analyses of both the grouped data from all ages of sampling and results from samples at specific ages. These results suggest that at a given birth weight in the range included in this study, infants with a greater MUAC have higher circulating T-cell counts in infancy.The results provide little evidence that nutritional status in infancy affects circulating leucocyte, lymphocyte and lymphocyte subpopulation counts. The only significant association observed was a strong positive correlation between infant plasma zinc status and absolute CD19+ counts. The interpretation of this finding within the current data is limited, but could support a role for zinc in the early development of B lymphocytes. However, given the almost universally accepted view that malnutrition impairs immune function, the general lack of association between the available measures on nutrient status and leukocyte counts merits some reflection . Numerouin utero. The significance of these findings is strengthened by the fact that analyses were carried out in relation to a single pre hoc definition of season, selected in the light of previous data relating birth season to adult mortality in this community [In conclusion, against a background of defined nutritional patterns and consistent individual immunological phenotype, we have shown an effect of birth season on circulating leucocyte, lymphocyte and lymphocyte subpopulation counts, with evidence for a seasonal effect ommunity . We founThe relevance of these early seasonal effects on lymphocyte subpopulation counts to the putative programming of long-term immunity remains to be established. Measures of functional immunity in the same cohort  and in aOne hundred and thirty-eight infants were recruited antenatally representing 94% of all singleton live-births within five participating West Kiang villages. The climate, economy, and predominant ethnicity of West Kiang are typical of rural Gambia as a whole. Multiple births and infants with birth weight below 1.5 kg, gestation less than 34 completed weeks, or major congenital malformation, were excluded. Of 184 identified pregnancies, 9 (5%) resulted in spontaneous abortion, 24 (13%) were multiple pregnancies or delivered outside the district, and the fates of a further 2 were unknown. Of the remaining 149 infants, 10 mothers (7%) declined to participate and one infant with congenital hydrocephalus was excluded. Seventy-five infants were male. At the time of the present study, HIV prevalence was estimated at 1.7% in reproductive Gambian women . None ofThe study was approved by the Joint Gambian Government/Medical Research Council Ethical Committee. Subjects were recruited with the informed consent of mothers following an explanation of the study in their own language.Birth-weight and gestational age (by Dubowitz score ) were meGrowth and morbidity were assessed fortnightly throughout infancy. Cord blood was collected at birth (10 ml) and venous samples obtained at 8, 16 and 52 week of age . Capillary blood was examined by thick film for malaria parasites every four weeks, in any infant with intercurrent fever of 37.5°C or above, and on each venous blood sample. Plasma alpha-1 antichymotrypsin (ACT) was also measured at 8 weeks and 52 weeks as an indicator of systemic acute-phase inflammatory responses. Infant feeding status was coded at each visit.T-lymphocyte , and natural killer cell (CD56+) subpopulations were enumerated in whole blood samples at birth (cord blood) and at 8, 16 and 52 weeks of age by flow cytometry  using appropriate isotypic controls . EDTA anticoagulated whole blood samples were labelled within 6 hours of collection, red cells were lysed according to the manufacturer's instructions, and the labelled cells stabilised and fixed. Stabilised samples were maintained at 4–8 degrees centigrade and analysed within 7 days at the Medical Research Council Laboratories, Fajara, the Gambia.Location of the lymphocyte gate, and freedom from monocyte contamination, were assessed using monoclonal antibodies to CD14+ and CD45+. The sum of CD4+8+19+56 was calculated as a second internal control, with values between 95% and 105% accepted as confirming acceptable purity ,29. AbsoPlasma ACT and micronutrient levels were measured at MRC Human Nutrition Research in Cambridge, UK. Plasma ACT and vitamin C levels assay on a Cobas-Bio centrifugal analyser . Prior to freezing, plasma samples for vitamin C analysis were mixed with an equal volume of 10% metaphosphoric acid in order to deproteinize the plasma. Plasma zinc concentrations were determined colorimetrically using a commercial kit . Plasma levels of retinol and carotenoids were measured by high-pressure liquid chromatography (HPLC).All data were analysed using Data Desk Version 6 . Population distributions were log-transformed prior to analysis. Effects on cell counts were assessed at each discrete age of sampling, and in datasets of samples at all ages combined, adjusted for days postnatal age. Cell counts were subjected to analysis of variance against defined exposure measures and potential confounders. Chi-squared and t-tests were used where appropriate.Exposure effects were analysed with and without adjustment for gender and gestation. Birth weight has previously been shown to be affected by birth season in this community . Therefopre-hoc hypotheses, and in the light of corroborative or contradictory evidence from the rest of the data .Outcome associations were considered significant at a level of p < 0.01. Results were further scrutinised for consistency of the association with the The outcome analyses at birth and 52-weeks of age were performed twice: first on the complete data sets, and then on subsets of data that excluded cord blood samples below 90% CD45+ cells in the lymphocyte gate (n = 7), and 52-week samples below 92% (n = 8). Descriptive statistics for the birth and 52-week samples were similarly calculated after excluding samples beyond these minimum levels of purity.ACC, SEM and AMP conceived and initiated the study. ACC conducted all fieldwork and data collection in The Gambia, conducted the statistical analyses and wrote the preliminary draft of the manuscript. PTN was responsible for all flow cytometric analyses. All authors contributed to and approved the final version of the manuscript. None of the authors had a conflict of interest to report.


Ptrend<0.01. Using hospital-based controls, the OR for 4000 g or more, compared to 2500–2999 g, was 1.30 (95% CI: 1.02–1.43) after adjusting for confounders . The results suggest that high birth weight is associated with increased risk of IL.In this paper, we compared the birth weight distribution among 201 infant leukaemia (IL) cases with that of 440 noncancer controls enrolled in Brazil in 1999–2005. Compared with the general population and the stratum 2500–2999 g as reference, IL cases weighing 3000–3999 g presented an odds ratio (OR) of 1.68 (95% CI: 1.03–2.76), and those of 4000 g or more, an OR of 2.28 (95% CI: 1.08–4.75),  MLL gene with other genes  with The BCSGIAL is a cooperative study group in different states of Brazil running a hospital-based case–control study to evaluate associations with IL maternal exposures to environmental risk factors during pregnancy. Details regarding case definition, control selection, exclusion criteria and data collection have been published previously , and their 95% confidence intervals, were ascertained for the birth weight strata  adjusted for selected confounders . The available national data on birth weight merged 3000–3499 g and 3500–3999 g strata, and this was followed with IL birth weight to allow the relevant comparisons. Statistical analyses were performed using unconditional logistic regression models, as described elsewhere , were enrolled in 1999–2005. The numbers of the main clinical and laboratory data were same as described previously  and 440 controls, stratified by age at diagnosis (0–12 P=0.0002); median birth weights were 3310 and 3226 g for ALL and AML, respectively. Children with ALL had slightly higher birth weights than AML, while male infants were slightly heavier than female infants. The birth weight for male IL infants was 3349.2 g (s.d.=49.6), with 95% CI: 3250.9–3447.5; for female IL infants, the birth weight was 3262.8 g (s.d.=48.8), 95% CI: 3166.0–3359.7. Further stratification by age revealed minimal differences between the strata before or after 12 months of age at diagnosis of IL (data not shown). It should, however, be remarked that MLL status in the current study was unknown for a larger proportion of children with low birth weight.Low birth weight (<2500 g) was found in 16 (8.0%) cases and in 74 (16.8%) controls. The comparison between IL and controls birth weight distributions showed that both ALL and AML weighed on average 3488 g at birth, compared with 3226 g among controls . Mean birth weight in MLL+ve cases (n=68) was 3340 g (s.d.=64.85), while in MLL−ve cases (n=80), it was 3310 g (s.d.=51.2); the medians were 3310 and 3300 g, respectively.Data were stratified according to P<0.01, MLL+ve showed a higher birth weight than MLL−ve cases at ages 12–21 months, and an inverse relationship in the first year of life was observed (figures available online).The magnitude of the association with birth weight IL was also explored in increasing weight strata  and 2. CDespite the many studies on childhood leukaemia and birth weight, few have focused on IL exclusively (P=0.001)  than IL cases (8.0%), which may have estimates of the magnitude of the association with birth weight . CompariP<0.01). A quite similar trend with a borderline statistical significance was also observed for MLL+ve cases (P=0.059).Comparisons with the general population revealed an insignificantly increased trend in ORs between ALL and birth weight  comparedve cases ; P=0.059Birth weight strata of hospital controls suggested a dose–response effect for IL, and also for ALL and AML subtypes after adjusting for certain confounders. For AML, discordant results have been reported with birth weight  promotes a proliferative advantage to damaged cells . The lin


To assess the effect of weight gain in childhood on blood lipid levels in adolescence.A population-based birth cohort carried out in Pelotas, Southern Brazil. All newborns in the city's hospitals were enrolled in 1982. The subjects have been followed up for several times in childhood. At age 18, 79% of all males were followed, and 2083 blood samples were available. Adjusted analyses controlled for household assets index, family income, parental schooling at birth, maternal smoking during pregnancy and breastfeeding duration.Birth weight for gestational age and weight gain in the first 20 months was not associated with blood lipid levels in adolescence. On the other hand, those subjects whose weight gain from 20 to 42 months of age was faster than that predicted from birth weight and weight-for-age z-score at the mean age of 20 months had lower high-density lipoprotein cholesterol (HDL) cholesterol [−0.78 ] and higher very low-density lipoprotein cholesterol (VLDL) and low-density lipoprotein cholesterol (LDL)/HDL ratio in adolescence. After controlling for current body mass index (BMI), the regression coefficient for HDL cholesterol decreased from −0.78 mg/dL to −0.29 mg/dL .Weight gain from 2 to 4 years is related to an atherogenic lipid profile in adolescence and this association is mediated by current BMI. Intrauterine malnutrition and early life growth patterns may result in metabolic and physiological programming with lifelong effects on the risk of selected diseases, including several cardiovascular conditions . Raised Huxley et al.  reviewedWeight gain up to adulthood has been positively associated with total cholesterol or triglycerides or negatively with high-density lipoprotein cholesterol (HDL) cholesterol ,6,7. On Studies in upper-income countries have observed that small for gestational age (SGA) infants often show rapid growth in the first two years . In low 2).The study was carried out in Pelotas (current urban population 320 000) in Southern Brazil. The population is mostly white, of Southern European descent. Like Brazil as a whole, there are wide social inequalities in health in this population. The infant mortality rate in the birth cohort was 38 per thousand live births and at the age of 18 years, 17.4% of all male subjects were overweight  were recruited and have been followed up on several occasions . There wIn 1984 and 1986, city censuses of over 70 000 households were carried out in search of children born in 1982; 87 and 84% of the original cohort were located, respectively, at the average ages of 20 and 42 months . MothersIn 2000, all males in the birth cohort were legally required to enlist in the Army. A research assistant was deployed at the recruitment office to interview all draftees and to verify if they belonged to the cohort. The cohort members were then invited to answer a questionnaire and to donate a blood sample. Typically, conscripts had continental-style breakfast at home at around 5:30 am, because they had to arrive at 6:00 am at the Army Base where the exams were carried out. Blood samples were collected by venepuncture between 10:30 am and 12:00 noon, after the whole round of medical, physical and intellectual exams had been completed. Male adolescents who did not attend the Army examination were sought at their last known address and invited to attend an examination at a clinic. Those who still failed to attend were visited at home. Overall, 79% of the cohort subjects were traced at this age. The study was approved by the Ethical Review Board of the Faculty of Medicine of the Federal University of Pelotas, and written informed consent was obtained from participating subjects.The following outcome variables were studied: total serum cholesterol; very low-density lipoprotein cholesterol (VLDL); LDL; HDL and LDL/HDL ratio. Total and HDL cholesterol, and triglycerides were measured using enzyme methods . LDL cholesterol was estimated using the Friedewald's formula  VLDL choBecause the ages of children seen at a given follow-up visit varied slightly, we used standard deviation scores (SDS). The SDS indicates the number of standard deviation of a measurement from the mean for that age and sex. Birth weight for gestational age z-scores was also computed. As Williams et al.  did not The following variables collected in the early phases of the study were considered as potential confounders:family income at delivery: total income earned by family members during the month before the interview;household assets index  ;parental schooling at delivery: years of schooling completed with success;breastfeeding duration: This variable was collected in 1984 and 1986. The earliest available information on breastfeeding cessation was used in order to minimize recall bias.maternal smoking during pregnancy .Blood lipids were analysed as a continuous outcome. The analyses took into account the correlation between weight gain in subsequent age ranges, as well as regression to the mean, by using conditional growth modelling . First, A total of 2250 subjects were interviewed at the age of 18 years, and 2083 blood samples were available. Added to the 143 cohort males known to have died, they represented 78.9% of all live born boys. Boys who were followed up in the army study had prevalence of low birth weight, preterm birth and being SGA age of 5.9%, 5.2% and 15.0%, respectively. Information on gestational age was not available for 429 (19.1%) of the subjects. The prospective nature of the study, its population basis, the use of standardized methods for birth measurements, gestational age assessment and anthropometric evaluation in infancy reduce the likelihood of selection and information bias. Furthermore, the subjects have been followed since birth, conversely to the majority of the studies that have relied on retrospective information on birth data , and lacOn the other hand, we were not able to obtain information on gestational age, for 19.1% of the subjects who consented to donate blood, and these subjects were, therefore, excluded from the analyses. Maternal recall of the date of the last menstrual period was lower among low birth weight infants, if maternal recall was also associated with blood lipid levels, the effect of birth weight for gestational age z-score could have been biased. Because blood lipids at 18 years of age were independent of maternal recall of the data of the last menstrual period, we do not believe that the above results have been due to selection bias.Other studies have assessed the effect of linear growth on blood lipids ,10,11. UWeight gain from 20 to 42 months was associated with decreased HDL cholesterol, as well as higher VLDL cholesterol. The relationship between rapid weight gain in later childhood and blood lipids was mainly explained by body composition in late adolescence. For example, after controlling for current BMI, the effect of weighing at 42 months one z-score more than that predicted from birth weight and weight at 20 months on HDL cholesterol decreased from −0.78 mg/dL to −0.29 mg/dL . Since the magnitude of the effect of weight gain in childhood was strongly reduced after controlling to current BMI, our data suggest that the effect of weight gain in later childhood on blood lipids is mediated by BMI in adolescence. Surprisingly, in another analysis, which included those subjects with information on weight at 20, 42 months and 15 years, weight gain from 42 months to 15 years was not related to blood lipids, in spite of its association with increased fatness in late adolescence .Previous analyses of our cohort show that catch-up growth in early infancy has short-term benefits on morbidity and mortality , and on


Leukemia is the most common childhood cancer. With the exception of Down syndrome, prenatal radiation exposure, and higher birth weight, particularly for acute lymphoid leukemia (ALL), few risk factors have been firmly established. Translocations present in neonatal blood spots and the young age peak of diagnosis suggest that early-life factors are involved in childhood leukemia etiology.We investigated the association between birth characteristics and childhood leukemia through linkage of the Minnesota birth and cancer registries using a case-cohort study design. Cases included 560 children with ALL and 87 with acute myeloid leukemia (AML) diagnoses from 28 days to 14 years. The comparison group was comprised of 8,750 individuals selected through random sampling of the birth cohort from 1976–2004. Cox proportional hazards regression specific for case-cohort studies was used to compute hazard ratios (HR) and 95% confidence intervals (CIs).Male sex , white race , and maternal birth interval ≥ 3 years  increased ALL risk, while maternal age increased AML risk . Higher birth weights (>3798 grams) , and one minute Apgar scores ≤ 7  increased risk for both types of leukemia. Sex was not a significant modifier of the association between ALL and other covariates, with the exception of maternal education.We confirmed known risk factors for ALL: male sex, high birth weight, and white race. We have also provided data that supports an increased risk for AML following higher birth weights, and demonstrated an association with low Apgar scores. Leukemia is the most common cancer occurring in children under age 15 in the United States with an annual incidence of 45.5 cases per million . TransloBirth characteristics and risk of childhood leukemia have been the focus of many studies and have been addressed in several reviews ,4,5. VarThe objective of this study was to examine the relation between characteristics recorded in birth records and childhood leukemia using a case-cohort study design which is relatively resistant to recall and selection biases that are of concern in case-control studies of childhood leukemia that frequently use survey based data collection methods.All childhood cancer cases diagnosed from 1988–2004 between the ages of 28 days and 14 years were ascertained from the Minnesota Cancer Surveillance System (MCSS) and linked to their birth records through probabilistic record linkage . It is eChildhood leukemia case groups were defined according to the International Classification of Childhood Cancer (ICCC) code  and inclBirth certificate variables that were examined included: parental sociodemographic characteristics: parental age, parental race, maternal birthplace, education, marital status; maternal reproductive characteristics: plurality, previous live births, prior fetal loss, last fetal death and live birth intervals; index pregnancy conditions: anemia, diabetes , hypertension , adequacy of prenatal care index ; labor aSAS version 9.1  was used for all statistical analyses. A modification of the stratified Cox proportional hazards model was used to compute risk estimates using a macro written by Langholz and Jiao  that comVariables were selected for multivariate models using a likelihood-ratio (LR) chi-square p-value cut-off of 0.1 for ALL comparing the intercept only model to that containing the covariate. For AML, only results from birth year stratified models are reported due to the small number of AML cases. Heterogeneity in ALL risk factors by sex or age group (<5 vs. 5–14 years) were examined by including an interaction term between sex or age group and the variable of interest. Tests of statistical significance for heterogeneity of risk were conducted using the test statement in PROC PHREG using a statistical criterion of p < 0.05.A total of 831 cases of leukemia occurring in children diagnosed from 28 days through 14 years were ascertained by MCSS from 1988–2004. Of these, 695 (84%) were matched to birth records including 12 leukemia cases with Down syndrome  relative to those in the 5th-25th percentile (2496–3117 grams) were at increased risk for ALL and AML. A pattern consistent with a linear relation between birth weight and ALL was observed for all percentile categories above the reference with the highest risk for individuals above the 95th percentile . Large size for gestational age , and low one minute Apgar scores  increased the risk for both leukemia subtypes. Down syndrome was associated with an increased risk for ALL of 19.8  and for AML of approximately 300-fold  (data not shown), similar to what has been previously reported [Several notable associations were found between ALL and AML and infant characteristics Table . Male sereported .No statistically significant differences in risk between sexes were noted for any parental or infant characteristics for ALL with the exception of maternal education (p = 0.04) in birth year adjusted models, which was associated with an approximately thirty percent increased risk for males whose mothers had education beyond high school but not females (data not shown).We conducted multivariate analyses for cases diagnosed at 0–14 years of age Table  and by aThe results of this study confirm previously established risk factors for ALL: male sex, white race, and higher birth weights. Further, children whose siblings were born three or more years prior were at an increased risk for ALL, while those who had low one minute Apgar scores were at an increased risk for both leukemia subtypes. Risks for ALL by age group and sex were generally similar with the exception of that for maternal education in males.Mothers of AML cases reported being foreign born more often than those of sub-cohort members. However, there was no distinct pattern in maternal nationality suggesting that this association may be a chance finding. AML cases were also more likely to have received inadequate prenatal care. However, a biological rationale for this association is lacking leading us to conclude that random variation is a more plausible explanation.Mothers of AML cases were also more likely to have gained ≥30 pounds during their pregnancy in contrast to another study that showed no increased risk for AML associated with maternal gestational weight gain in excess of thirty pounds . It is oAML risk increased linearly with maternal age in birth year adjusted analyses. Of note is that maternal age also had a significant linear effect on ALL risk in 0–4 year olds in birth year adjusted models (data not shown). The relation between advanced maternal age (defined as ≥35 in most studies) and leukemia has been examined in many studies with inconsistent findings. Studies reviewed by Little published prior to 1998 and those since for leukemia overall or ALL also support a slight positive association with a mean risk estimate of around 1.2 for mothers who were ≥35 years at the time of birth compared to younger mothers ,15-25. MIndividuals who were ≥ 3 years younger than their previous live-born sibling were at increased risk for ALL relative to children whose mothers had no other live births. Longer birth intervals have been associated with adverse maternal outcomes, particularly labor complications and pre-eclampsia . In our Substantial evidence indicates that high birth weight (≥ 4000 grams) is a risk factor for childhood leukemia, particularly ALL . Our resLow one minute Apgar scores increased the risk for both AML and ALL. Apgar scores are used to rapidly evaluate the general health of the infant shortly after delivery with values above 7 considered excellent while values below 7 can indicate an infant who has undergone oxygen deprivation and is at increased risk of neonatal death . BecauseOne of the strongest and yet unexplained risk factors for childhood ALL is male sex. We did not find any statistical differences in risk factors by sex with the exception of maternal education which seems unlikely to be due to anything more than chance. Few studies have reported sex-specific risk estimates. Three studies that examined the birth weight association in sex-specific analyses have reported divergent results with one study showing that the increased risk for higher birth weights was limited to males  and two A major strength of our study is the case-cohort study design and collection of exposure data prior to disease onset obviating recall bias of particular concern in case-control studies frequently collecting data through parental interview after disease status is known. Retrospective case-control studies are also susceptible to selection bias, where estimates could be biased if the exposure distribution of the comparison population is not representative of the source cohort. Because we randomly sampled members of the sub-cohort from the entire birth cohort distribution of cases, selection bias was avoided.Our study also has limitations. We did not have follow-up information on non-case members of the sub-cohort or have knowledge of cases diagnosed prior to 1988 (the year of inception of the cancer registry). It is also possible that some sub-cohort members who were later diagnosed with leukemia were not captured by MCSS due to out-migration. However, given the rarity of childhood leukemia, it is unlikely that out-migration or missed sub-cohort diagnoses prior to 1988 could have caused a large bias in our results. Also of note is that the success of matching cases captured by MCSS to birth records decreased with diagnosis age. The most likely reason for this is due to immigration of individuals born elsewhere which would not have biased our results because our study was of Minnesota born children. A number of individuals had missing data for variables recorded in birth records. Biased estimates from missing data occur when the data are not missing at random . We do nWe have reported findings from a population-based case-cohort study of childhood leukemia in Minnesota. Our study adds to the current knowledge of leukemia risk factors, particularly birth weight with evidence supporting a positive linear relation between birth weight and both ALL and AML. Our data also suggests that early-life factors recorded in birth records do not explain the male excess found in ALL. Further, we have shown an increased risk between low Apgar scores and leukemia, the reasons for which should be the subject of further investigation.The author(s) declare that they have no competing interests.KJ conducted the analysis and prepared the paper for editing by coauthors and submission. JS assembled the data for analyses, retrieved additional data elements from the cancer and birth registries, and reviewed and edited the manuscript. SP provided statistical assistance and reviewed and edited the manuscript. JR & LS reviewed the proposal for the manuscript, all analyses, and edited manuscript drafts.The pre-publication history for this paper can be accessed here:


Parental investment is a limited resource for which offspring compete in order to increase their own survival and reproductive success. However, parents might be selected to influence the outcome of sibling competition through differential investment. While evidence for this is widespread in egg-laying species, whether or not this may also be the case in viviparous species is more difficult to determine. We use pre-industrial Finns as our model system and an equal investment model as our null hypothesis, which predicts that  middleborns should be disadvantaged through competition. We found no overall evidence to suggest that middleborns in a family are disadvantaged in terms of their survival, age at first reproduction or lifetime reproductive success. However, when considering birth-order only among same-sexed siblings, first-, middle- and lastborn sons significantly differed in the number of offspring they were able to rear to adulthood, although there was no similar effect among females. Middleborn sons appeared to produce significantly less offspring than first- or lastborn sons, but they did not significantly differ from lastborn sons in the number of offspring reared to adulthood. Our results thus show that taking sex differences into account is important when modelling birth-order effects. We found clear evidence of firstborn sons being advantaged over other sons in the family, and over firstborn daughters. Therefore, our results suggest that parents invest differentially in their offspring in order to both preferentially favour particular offspring or reduce offspring inequalities arising from sibling competition. In species with altricial young, parental investment has profound effects on offspring survival et al. provides a useful null hypothesis against which parental investment tactics can be measured et al. will not be upheld. For example, if parental experience is important for offspring success, or if parental investment increases as expected residual reproductive value decreases et al. , many have failed to control for many of the potential confounding terms; which might partly account for contradictory results (see above). It thus would appear critical to control for all the factors that are known to be linked both to birth order and to survival or reproductive success. Some studies, for example, controlled for socio-economic status, but failed to control for total number of siblings or maternal age i.e. lifetime fecundity (number of offspring born) and offspring survival probability. Finally, we investigate the effects of being first-, middle- or last-born on the underlying life-history traits that influence lifetime reproductive success: age at first and last reproduction and inter-birth intervals. Although the equal parental investment model . During the study period, these populations depended on farming for their livelihood, supplemented with fishing in the archipelago areas of south-western Finland. Farmers could either own their land, or rent it (tenant farmers). Information on the occupation of each man, or for women, that of their husband, allowed us to control for socio-economic status. We contrasted two groups, according to those owning land versus those either renting or having no access to land at all. Importantly, these categories have already been associated with variation in individual life-history traits including fitness among humans living in pre-industrial conditions The study sample includes 2180 men and 2126 women (F1 generation) born between 1732 and 1882 to 716 mothers (P generation). We recorded full life-history data  for all F1 individuals, and followed their offspring (F2) until age 15. Birth-order from the mother's point of view was used to build the variable of interest, comprising three categories: firstborns, middleborns and lastborns. Twins were excluded from the sample, as were families of less than three children because the middleborn resource handicap can only be considered in families of three or more children  [22]. Thvs. landless); geographic area (mainland vs. archipelago); birth cohort (5 categories with 20 years blocks). In all models, the identity of mothers was fitted as a random term to account for the use of several offspring within families. Statistically significant terms at the level of 0.05 were determined. Once the minimal model was found, birth-order category (our term of interest) was added and its significance determined. All biologically meaningful two-way interactions involving birth-order  were also tested but only included and reported here if statistically significant.Statistical analyses used to address our questions were conducted using SAS . In all analyses, the following variables were entered into a model to control for potentially confounding sources of variation: sex; mother's age at birth; mother's survival (age at mother's death); number of siblings; parental socio-economic status  were analysed with Generalised Linear Mixed Model (GLMM) where the response term was fitted to a binomial error structure and a logit link function (using GENMOD function in SAS). For binary response terms, binomial denominator was fixed at 1. For the proportion of offspring surviving to adulthood, the number of surviving offspring (LRS) was considered as the response term with a variable binomial denominator equal to the number of offspring born (fecundity). The count response variables (LRS and fecundity) were analysed with GLMM where the response term was fitted to a Poisson error structure and a logarithm link function (using GENMOD function in SAS). The continuous response variables  were analysed with linear mixed-effects models  with the response terms being fitted to a normal error structure and Satterthwaite's formula being used to approximate the denominator degrees of freedom of each fixed effect To determine whether birth-order affected survival probability to adulthood (age 15), we included all individuals who died before age 15 or were successfully followed at least until age 15 . For the probability of reproducing, we restricted our analyses to individuals who survived to age 15 and who had been successfully followed at least until the age at which 90% of the individuals in the population had already finished reproducing, if they were to ever reproduce in their lifetime (age 50 for men and 44 for women). If the individual disappeared after this age and had not reproduced before, he/she was considered as having never reproduced. The subset of individuals for which reproductive events were known comprised 843 men and 913 women . These individuals did not differ from those excluded from the sample in the proportions of first-, middle- and lastborns .LRS was measured as the lifetime number of F2 offspring raised to 15 years. We restricted the sample to include only those individuals who had been successfully followed at least until the age at which 90% of the individuals in the population had already finished reproducing (age 50 for men and 44 for women), for which LRS is known, and who reproduced at least once (739 men and 794 women).Two life-history traits will govern our measure of LRS: lifetime fecundity (number of offspring born) and the survivorship of offspring to adulthood. Lifetime fecundity was considered as a count response term. The proportion of offspring surviving to age 15 was examined by considering LRS as a response term in a GLMM with logit link function and a variable binomial denominator equal to fecundity. For these analyses, we used the same sample as for LRS.i.e. length of reproductive life divided by the number of children). For the analysis of inter-birth interval, we used the sub-sample of individuals who reproduced at least twice (694 men and 731 women). Finally, among men who survived to adulthood, we investigated whether being a first-, middle- or lastborn son influenced socio-economic status, i.e. whether they owned a land. The sample of families with at least three sons, for which both own and parental socio-economic status were known, comprised 947 men.We examined more specific life-history traits involved in LRS: age at first reproduction, age at last reproduction, and average inter-birth interval , and negatively associated with the number of siblings . After controlling for these significant effects, we found clear, although marginally non-significant, trend for differences between overall birth-order categories  in their probability of surviving to adulthood, with firstborns being disadvantaged, and lastborns having the best chances of surviving . Survival was positively associated with mother's (P generation) survival , but we found that firstborn daughters had a significantly reduced probability of surviving to adulthood, as compared to their sisters  and for mother's survival . The model explained 12% of the variance.When considering intra-sex birth-order, we did not find any significant differences between first-, middle- and lastborn sons in their probability of surviving to adulthood . 22 = 1.39, P = 0.5), or among brothers only , or among sisters only .We found no evidence to suggest that probability of reproducing in a lifetime depended on birth-order category among all siblings , and between cohorts . In this subset of individuals who did reproduce, men had more children than women . LRS was positively associated with P mother's survival . After controlling for these effects, we found no evidence to suggest that LRS depended on birth-order category among all siblings .Among F1 individuals who survived to age 15, who were successfully followed until the end of their potential reproductive life, and who reproduced at least once, the average LRS was 3.3±2.2 s.d., and the maximum was 11. LRS differed between geographical areas . However, first-, middle- and lastborn sons did differ , controlling for total number of siblings : firstborn sons were strongly advantaged  and between birth-cohorts . In this subset of individuals who did reproduce, men had more children than women . Lifetime fecundity was also positively associated with P mother's survival  and with the number of siblings . After controlling for these effects, we found no evidence to suggest that lifetime fecundity was associated with overall birth-order category .Among F1 individuals who survived to age 15, who were successfully followed until the end of their potential reproductive life, and who reproduced at least once, the average lifetime fecundity (number of children born) was in our sample 5.4±3.0 s.d.  and the maximum was 16. Lifetime fecundity differed between geographical areas , total number of siblings  and mother's survival . This model explained 17% of the variance. Additionally, we found that when each adult son's own socio-economic status was included in the model , this effect of intra-sex birth-order category on their fecundity was still significant , suggesting that that the favoured position of first-borns sons as the inheritors of the parental land was not the sole reason for their higher lifetime fecundity. No differences were found between sisters' birth order categories for fecundity .However, when investigating the effect of intra-sex birth-order categories, we found that firstborn sons produced significantly more offspring than their brothers and that middleborn sons produced the smallest number of offspring , between birth-cohorts , and increased with socio-economic status . In addition, it was negatively associated with the number of siblings . After controlling for these effects, we found no evidence to suggest that the proportion of an individual's offspring surviving to adulthood differed according to his/her overall birth-order categories in a family  or intra-sex birth-order categories .On average, 62% of offspring (F2) survived to adulthood . This percentage differed between geographical areas  and was positively associated with maternal age (P generation) . After controlling for these effects, we found that age at first reproduction did not significantly depend on overall birth-order category . However, when considering intra-sex birth-order categories, we did find that firstborn sons began reproduction on average almost two years earlier than their brothers .The average age at first reproduction among F1 individuals differed between men and women  and was positively associated with mothers' survival . After controlling for these effects, we found no evidence to suggest that age at last reproduction depended on overall birth-order categories  or intra-sex birth-order categories .The average age at last reproduction differed between men and women (Fi.e. the length of reproductive life divided by the number of children produced) differed between birth cohorts  and depended on parental socio-economic status , but we found no difference between overall birth-order categories in a family .Finally, among individuals who reproduced at least twice, average inter-birth interval . After controlling for this effect, we found that intra-sex birth-order category also had a significant effect, with firstborn sons being much more likely than their brothers to become landowners (χ22 = 26.47, p<0.0001; Among men, own socio-economic status as an adult was strongly positively associated with parental socio-economic status (χParental investment has profound effects on offspring success et al. posit that all else being equal, differences in the amount and quality of parental investment received by offspring can arise even in the presence of an equity motive on the parents' side: in families of three or more children, middleborns will receive fewer resources due to competition with younger and elder siblings Hertwig We found some evidence that parental inexperience negatively influenced offspring survival to adulthood. Children who had the lowest chances of surviving were firstborns. This was especially true for firstborn daughters, who were less likely to survive than their sisters. Firstborns are often born smaller in humans et al.In fact, modelling resource allocation among different-aged offspring indicates that, in nearly all circumstances, the evolutionary stable strategy is to bias investment toward older offspring et al. model is that it fails to consider the possibility that older brothers and older sisters are likely to have, at least in some socio-ecological settings, a different influence on their younger siblings' fitness. For example, in rural Gambia, children with any living sisters who were at least 10 years older than the child had lower mortality rates, whereas the presence of older brothers did not improve survival chances th century Sweden. Finally, using the same dataset as the present study, Rickard et al.A further limitation of the Hertwig et al. model fails to take into account differences in the timing of resource acquisition. In the ‘equal investment’ model, the resources received for the first and last birth rank are identical in quantity, but not in timing. For example, consideration of the resources received during the first two years of first- and lastborns in a family of four children (Finally, the Hertwig children , illustrIn conclusion, we have shown that birth-order category in the sibship as a whole has little influence on fitness. By contrast, we found that, among male offspring in this population, firstborn sons had a significantly higher fitness than their brothers, and middleborn sons tended to have the lowest fitness. These results contrast with predictions of the ‘equal investment’ model


Jane Norman and colleagues analyzed linked perinatal surveillance data in Scotland and find that between 1980 and 2004 increases in spontaneous and medically induced preterm births contributed equally to the rising rate of preterm births. Rates of preterm birth are rising worldwide. Studies from the United States and Latin America suggest that much of this rise relates to increased rates of medically indicated preterm birth. In contrast, European and Australian data suggest that increases in spontaneous preterm labour also play a role. We aimed, in a population-based database of 5 million people, to determine the temporal trends and obstetric antecedents of singleton preterm birth and its associated neonatal mortality and morbidity for the period 1980–2004.p<0.01) and medically indicated preterm births , which persisted when adjusted for maternal age at delivery. The greater proportion of spontaneous preterm births meant that the absolute increase in rates of preterm birth in each category were similar. Of specific maternal complications, essential and pregnancy-induced hypertension, pre-eclampsia, and placenta praevia played a decreasing role in preterm birth over the study period, with gestational and pre-existing diabetes playing an increasing role. There was a decline in stillbirth, neonatal, and extended perinatal mortality associated with preterm birth at all gestation over the study period but an increase in the rate of prolonged hospital stay for the neonate. Neonatal mortality improved in all subgroups, regardless of obstetric antecedent of preterm birth or gestational age. In the 28 wk and greater gestational groups we found a reduction in stillbirths and extended perinatal mortality for medically induced but not spontaneous preterm births  although at the expense of a longer stay in neonatal intensive care. This improvement in stillbirth and neonatal mortality supports the decision making behind the 34% increase in elective/induced preterm birth in these women. Although improvements in neonatal outcomes overall are welcome, preterm birth still accounts for over 66% of singleton stillbirths, 65% of singleton neonatal deaths, and 67% of infants whose stay in the neonatal unit is “prolonged,” suggesting this condition remains a significant contributor to perinatal mortality and morbidity.There were 1.49 million births in Scotland over the study period, of which 5.8% were preterm. We found a percentage increase in crude rates of both spontaneous preterm birth per 1,000 singleton births . Nowadays in the US, for example, more than half a million babies arrive earlier than expected every year (1 in 8 babies). Although improvements in the care of newborn babies  mean that preterm babies are more likely to survive than in the past, preterm birth remains the single biggest cause of infant death in many developed countries, and many preterm babies who survive have long-term health problems and disabilities, particularly those born before 32 weeks of gestation. Preterm births can be spontaneous or medically induced. At present, it impossible to predict which mothers will spontaneously deliver early and there is no effective way to prevent these preterm births; medically induced early labor is undertaken when either the unborn baby or mother would be at risk if the pregnancy continued to full term.Preterm birth rates need to be reduced, but before this can be done it is important to know how the causes of preterm birth, the numbers of preterm stillbirths, and the numbers of preterm babies who die at birth  or soon after  are changing with time. If, for example, the rise in preterm births is mainly due to an increase in medically induced labor and if this change in practice has reduced neonatal deaths, it would be unwise to try to reduce the preterm birth rate by discouraging medically induced preterm births. So far, data from the US and Latin America suggest that the increase in preterm births in these countries is solely due to increased rates of medically induced preterm births. However, in Europe and Australia, the rate of spontaneous preterm births also seems to be increasing. In this study, the researchers examine the trends over time and causes of preterm birth and of neonatal death and illness in Scotland over a 25-year period.By searching a Scottish database of linked maternity records and infant health and death records, the researchers identified 1.49 million singleton births that occurred between 1980 and 2004 of which nearly 90,000 were preterm births. Over the study period, the rates of spontaneous and of medically induced preterm births per 1,000 births increased by 10.7% and 41.2%, respectively, but because there were more spontaneous preterm births than medically induced preterm births, the absolute increase in the rates of each type of birth was similar. Several maternal complications including preeclampsia (a condition that causes high blood pressure) and placenta previa (covering of the opening of the cervix by the placenta) played a decreasing role in preterm births over the study period, whereas gestational and preexisting diabetes played an increasing role. Finally, there was a decline in stillbirths and in neonatal and perinatal deaths among preterm babies, although more babies remained in the hospital longer than 7 days after birth. More specifically, after 28 weeks of gestation, stillbirths and perinatal deaths decreased among medically induced preterm births but not among spontaneous preterm births.These findings indicate that in Scotland between 1980 and 2004, increases in spontaneous and medically induced preterm births contributed equally to the rising rate of preterm births. Importantly, they also show that the increase in induced preterm births helped to reduce stillbirths and neonatal and perinatal deaths, a finding that supports the criteria that clinicians currently use to decide whether to induce an early birth. Nevertheless, preterm births still account for two-thirds of all stillbirths, neonatal deaths, and extended neonatal stays in hospital and thus cause considerable suffering and greatly increase the workload in neonatal units. The rates of such births consequently need to be reduced and, for Scotland at least, ways will have to be found to reduce the rates of both spontaneous and induced preterm births to achieve this goal while continuing to identify those sick babies who need to be delivered early to give them the best chance of survival.http://dx.doi.org/10.1371/journal.pmed.1000153Please access these Web sites via the online version of this summary at Tommys is a nonprofit organization that funds research and provides information on the causes and prevention of miscarriage, premature birth, and stillbirthpreterm birth (in English and Spanish)The March of Dimes, a nonprofit organization for pregnancy and baby health, provides information on premature babies (in English and Spanish)The Nemours Foundation, another nonprofit organization for child health, also provides information on maternal and infant health (in English and Spanish)The US Centers for Disease Control and Prevention provides information on pregnancy, including a section on pregnancy complicationsThe US National Women's Health Information Center has detailed information about premature babies and to information on pregnancy (in English and Spanish)MedlinePlus provides links to other information on  Preterm birth is the pre-eminent problem facing perinatologists in developed countries and has been defined as a major public health problem whose magnitude is increasing Preterm birth may result from induced labour or operative delivery  in either the fetal or maternal interest  or it may follow spontaneous preterm labour with or without preterm premature ruptured membranes Recent reports from the US have suggested that much of the increase in singleton preterm birth rates has resulted from rising rates of medically indicated preterm birth, and that spontaneous preterm birth rates are static or falling, especially amongst high risk ethnic groups The aim of this study was to determine the temporal trends in obstetric antecedents of preterm birth, and the temporal trends in neonatal mortality and morbidity associated with each of these antecedents during the period 1980 to 2004, using Scotland's comprehensive perinatal database We explored the SMR02/SMR11/SBR/SSBID/GROS Birth Database . This database contains linked maternity, neonatal, and stillbirth/infant death records, with records pertaining to mother and baby held together. The SMR02 (Scottish Morbidity Records 2) return is completed at the time of discharge of any patient from a Scottish maternity hospital and the level of completeness over the period studied is estimated to be in excess of 98%. There is also a facility for the data to be returned in the case of home births but this is thought to be less complete. However, home births in Scotland during this period comprised less than 1% of all births, so this is unlikely to be a source of significant error. SMR11  are routine neonatal returns. SSBID is a record relating to Stillbirths and Infant Deaths, based on stillbirths and infant deaths that are registered with the General Register Office for Scotland (GROS). For each SSBID event, further information is sought from the relevant hospital. Because the data are based on registered events, it is unlikely that any cases are missed.We examined records on singleton births between 1975 and 2004. Although we initially planned to report on data throughout this period, after an initial review we opted to confine our analysis to the period 1980–2004, excluding data from 1975–1979. The rationale for this was 2-fold: firstly, the coding changed from ICD8 to ICD9 in 1979 and many of the relevant codes do not map satisfactorily between these two systems, and secondly, the overall data quality in the early years of SMR02 appeared not to be adequate enough for this study, with problems such as augmentation with oxytocin probably being erroneously recorded as induction in a number of records.We calculated the number and percentage of all singleton preterm births (live and stillbirths before 37 completed wk of gestation) and the number and percentage of singleton preterm births between 24 and 27 wk, 28 and 31 wk, and 32 and 36 wk. Gestational age at birth (as recorded on SMR02) is based on the clinician's best “guess” of estimated date of delivery, based largely on last menstrual period and ultrasound findings. The latter will have been used for confirmation in the majority of women—since the early 1990s, more than 95% of pregnant women in the UK (including Scotland) have had ultrasound in the first half of their pregnancy Thereafter, we calculated outcomes for 5 calendar year periods starting in 1980; i.e., from the period 1980–1984 up to 2000–2004. Initially preterm birth was classified according to its obstetric antecedents (see below for definition). These antecedents were used to subdivide preterm birth numbers and percentages. Outcomes were then further analysed by gestation at delivery .The obstetric antecedents of preterm birth were defined as follows: spontaneous preterm labour with maternal complications, spontaneous preterm labour without maternal complications, pPROM with maternal complications, pPROM without maternal complications, indicated preterm delivery (labour induction or elective caesarean delivery) with maternal complications, and indicated preterm delivery without maternal complications. These antecedents are similar to those used by other investigators in this field We calculated secular trends in rates of preterm birth associated with these maternal complications. The relative contribution of each of the maternal complications to preterm birth was calculated for the 10-y period 1995–2004. Additionally we determined the following neonatal outcomes associated with preterm birth: birthweight, stillbirth, neonatal death, extended perinatal mortality , and the incidence of prolonged stay in the neonatal unit (defined as more than 7 d) Permission for record linkage and analysis of data for the purpose of this study was obtained from the Privacy Advisory Committee of NHS National Services Scotland.Mantel-Haenszel chi-square (linear by linear association) test for trend was performed on temporal trends for the obstetric antecedents and subdivided by gestation groups to examine whether there was a linear relationship over the time period.p values for all hypothesis tests were two-sided, and we set the significance at p<0.05. The goodness of fit of the logistic regression models was assessed using the Hosmer and Lemeshow test. We used SPSS 13.0 software package to conduct all the statistical analyses.Univariate analysis was initially carried out to examine each individual confounding factor in relation to the outcomes of preterm birth, stillbirth, low birthweight, neonatal death, and prolonged stay in neonatal care. Thereafter, we used multivariate logistic regression modelling to examine the relationship between preterm birth, neonatal outcomes, and the six obstetric antecedents. All confounding factors including birth cohort period, deprivation, parity, smoking status (only recorded since 1993), health board of residence, age, and gestation at delivery (where appropriate) were entered into the model. The outcomes were considered as dichotomous variables and the covariates above categorised. The p<0.01) but greater percentage increase in medically indicated preterm birth rates  over the study period . Looking at gestational age subgroups, the rise in the rate of spontaneous preterm birth held for all gestation subgroups, whereas the relative increase in rates of elective/induced preterm delivery was greatest in the 24–27 wk gestation subgroup. There was a modest increase in rates of induced/elective preterm birth in the 32–36 wk gestation subgroup but a 22% decline in the 28–31 wk gestation subgroup. However, given the much larger contribution of the 32–36 wk gestation subgroup to overall numbers, there was a net increase of 34% in induced/elective deliveries in babies of gestational age 28–36 wk over the study period  were preterm (before 37 completed wk gestation). As previously shown, rates of preterm birth per 1,000 singleton births increased over the time period studied from 5.4% singleton live births in the 5-y period beginning 1980 to 6.3% in the 5-y period beginning 2000 y period . The incy period .In view of the increase in maternal age of first pregnancy over the study period  [16],) .Looking at secular trends, there was an increase in the incidence of prolonged stay in hospital in the period 1995–1999 compared with 1980–1984 . This isz score<2) were more common amongst preterm compared with term births  . The adverse effects of deprivation, smoking, and nulliparity on the incidence of SGA were similar to their effects on stillbirth and prolonged hospital stay. After adjustment for parity, deprivation, smoking, birth cohort period, and health board of residence, preterm birth was paradoxically associated with a reduction in odds of SGA. Looking at obstetric antecedents, in the absence of maternal complications, babies born as a result of induced/elective preterm birth were at an increased risk of being SGA after adjustment whereas those born preterm following pPROM were at reduced risk  , increased rates of idiopathic spontaneous labour (with or without pPROM) would be a major contributor to the increase in preterm birth rates observed in Scotland. Our hypotheses were partially correct—although the percentage increase in preterm birth rates was greatest in the elective/induced category (a 41% increase over the study period), rates of preterm birth were also rising in the spontaneous preterm birth category (a 10% rise over the study period). These changes persist when adjusted for maternal age: a progressive rise in the proportion of preterm births (in both the elective and spontaneous category) would have occurred even if there had been no change in maternal age over the study period. Although the percentage rise in elective/induced preterm births is greater than that in spontaneous preterm births (with and without pPROM), the absolute increase in the rate of preterm births is similar in each group . Thus, The findings that pregnancy induced hypertension and abruption are the commonest maternal complications preceding preterm birth are not unexpected. However, the decline in rates of preterm birth in association with essential hypertension, pregnancy induced hypertension, eclampsia, and placenta praevia over the study period was not anticipated. It could imply either better obstetric management of these conditions , it could imply a greater willingness of obstetricians to manage these conditions conservatively, or it could be an artefact of varying completeness of coding. In contrast, we observed a 4- to 7-fold increase in preterm deliveries associated with pre-existing and gestational diabetes. The reasons for the increased contribution of diabetes to preterm delivery are likely to be multifactorial: greater numbers of women with pre-existing (both Type I and Type II) diabetes may be getting pregnant, ascertainment of gestational diabetes may be getting better, and the true incidence of gestational diabetes may be rising as obesity rates rise in pregnancy. The fact that gestational diabetes increased the odds of spontaneous preterm labour was somewhat of a surprise, and has not (to our knowledge) been highlighted in previous studies. Whatever the reason for the association between pre-existing and gestational diabetes and preterm birth, it suggests that treatment and management of diabetes during pregnancy will be a key issue in inhibiting any further rise in preterm birth rates, particularly given that the maternal condition associated with the greatest odds of preterm birth is pre-existing diabetes.The reason for the increase in rates of induced/elective preterm birth in the absence of maternal complications is unclear. Induced/elective preterm delivery in this scenario implies detection of a fetal complication. However, there have been no new methods of fetal surveillance over this time period: indeed the current recommended tools for surveillance of the small-for-dates fetus are umbilical Doppler ultrasound and fetal biometry—both of which tools have been available for the majority of the study period In terms of spontaneous preterm labour in the absence of maternal complications, again there has been no systematic change in practice over the time period of the study that is likely to contribute to the increase in rates. Current UK guidelines suggest that it is reasonable not to use tocolytic agents given their lack of effect on outcomes or preterm delivery rates overall A caveat to any study of this type has to be that we are relying on routinely collected data for our analysis, with issues of data quality and of other confounders that we were unable to measure. Although we did not formally examine SMR2 data quality for the purposes of this project, previous studies have confirmed that it is adequate for analyses of the type reported here. For example, an evaluation over a 6-mo period in 1996–1997 compared a 5% sample of paper records with the database. This showed that most fields had less than 2% errors. However, there was a 5.6% error rate in the estimated gestation, 6.4% error in induction of labour, 13.5% error in duration of labour, and a 10%–20% error rate in the recording of the main ICD diagnostic codes The results of our study show that to reverse the rising trend of preterm birth in Scotland, policies and research strategies should be focussed on both spontaneous and elective/induced preterm deliveries: different approaches will be required, but both are important. They also emphasise an emerging theme in the epidemiology of preterm birth in that although preterm birth rates are rising in all developed countries, the rise appears to be driven largely by rises in elective/induced preterm birth in the US and South America, but by an increase in both elective/induced and spontaneous preterm births in Europe and Australia.The data supported our second hypothesis: that neonatal outcomes of preterm birth would show a progressive improvement over the time period studied. There was a 55% reduction in risk of neonatal death and a 10% reduction in risk of both stillbirth and extended perinatal death over the study period. However, these improved outcomes come at a cost to the health service with a 57% increase in the risk of prolonged stay in hospital for the neonate over the study period. These data are in keeping with those from other groups, showing improved perinatal outcomes in association with preterm birth over the time period of our study. The improvement in stillbirth neonatal mortality in the elective/induced preterm birth group at gestations of 28 wk and greater does support the decision making behind the rising trend of elective/induced preterm birth. It implies that sick babies are being appropriately identified, and rather than being stillborn or dying shortly after birth as they were in earlier time cohorts, they are delivered and survive. Whatever the specific indication for the elective/induced preterm delivery, the fact that elective/induced preterm babies are more likely than those delivered spontaneously to be SGA suggests that their in utero environment is compromised and has restricted their growth. However, once delivered alive, somewhat surprisingly these babies do better than those delivered spontaneously  in terms of lower neonatal mortality and a lower incidence of prolonged hospital stay.Our data showing lower neonatal mortality in electively delivered babies are somewhat different to those of Barros and Velez Mdel, who showed that babies born following medically indicated preterm birth, even in the absence of maternal complications, had higher neonatal mortality than those delivered following spontaneous preterm labour The rationale for poorer neonatal outcomes following spontaneous rather than elective preterm delivery may be that intrauterine infection, which often triggers spontaneous preterm labour, continues to have an adverse effect on the neonate after delivery. This likely lower neonatal mortality in electively delivered babies compared with those delivering spontaneously should be taken into account by obstetricians and paediatricians making decisions about elective preterm delivery of individual babies .To summarise the neonatal outcomes of this study, the secular trend is of a reduction in mortality associated with preterm birth, supporting the decision making behind elective/induced preterm delivery, particularly at gestations of 28 wk and above. Although babies born electively  tend to be smaller than those delivering spontaneously, their outcomes tend to be at least as good. In contrast, the rising tide of antepartum stillbirth, which increasingly appears to be a factor in triggering elective preterm birth , has been noted elsewhere, suggesting that research into this issue is needed. Lastly, a common theme throughout our data was that smoking and social deprivation both continue to play a significant role in the aetiology of preterm birth and are major risk factors for adverse neonatal outcomes.


The risk profile of childhood leukaemia in Greece was studied through a case-control investigation that included all 153 incident cases of the disease, ascertained throughout the country during 1993 and 1994, and two hospital controls for every case matched for gender, age and place of residence. The data were analysed using conditional logistic regression and the associations are expressed in terms of adjusted odds ratios (OR) and their 95% confidence intervals. Cases were born to mothers of a higher standard education, the OR for an increment of four schooling years being 1.48 (1.17-1.87) and had higher birth weight, the OR for an increment of 500g being 1.36 (1.04-1.77). Pet ownership and birth after a pregnancy with anaemia were associated with increased risk, the ORs being 2.18 (1.14-4.16) and 2.60 (1.39-4.86) respectively. From the frequency analyses, indicative inverse associations were found with birth order, household crowding and previous hospitalization with allergic diseases, whereas indicative positive associations were found with diabetes mellitus during pregnancy and with neonatal jaundice. Substantial or significant elevations were not found with respect to maternal smoking and coffee drinking during pregnancy, diagnostic radiography and ultrasonographic examinations or blood transfusions. A significant inverse association with maternal consumption of alcohol could be due to multiple comparisons, but a detrimental effect can probably be excluded. A non-significant positive association with total shots of viral vaccinations and a weak non-significant inverse association with breast feeding were also found. We interpret the findings of this study as being compatible with acute childhood leukaemia being linked with delayed development of herd immunity to fairly common infectious agents, in conjunction with accelerated perinatal and early post-natal growth.


P<0.01) after adjustment for gestational age and birth weight. For oestradiol, there was no difference in cord plasma levels between the severe pre-eclampsia group and controls, after adjustment for length of gestation and birth weight. These results suggest that an anti-oestrogenic effect associated with pre-eclampsia may be mediated through high levels of α-foetoprotein rather than low levels of oestradiol.Women born after pre-eclamptic pregnancies have been reported to be at reduced risk of breast cancer as adults, because of reduced intrauterine oestrogen influence on breast tissue; high levels of α-foetoprotein (a glycoprotein with anti-oestrogenic properties), however, could also be important. In severe pre-eclampsia, placental function and foetal growth are reduced, and umbilical cord plasma levels of α-foetoprotein could reflect the underlying processes. Umbilical cord blood was collected in 12 804 consecutive deliveries. Among 307 pregnancies with clinical pre-eclampsia, 66 singleton pregnancies were identified as clinically severe, and 610 singleton pregnancies were selected as controls. Oestradiol and α-foetoprotein were measured from umbilical plasma, and birth weight was standardized as the ratio between the observed and expected birth weight, adjusted for differences in gestation length and offspring sex. Cord plasma levels of α-foetoprotein were significantly higher in severe pre-eclampsia than controls  86, 728–731. DOI: 10.1038/sj/bjc/6600125www.bjcancer.comCancer Research UK© 2002  The Norwegian Medical Birth Registry records information on all deliveries that take place in the country (−1 (semiquantitative dipstick 1+) in at least one urine sample after 20 weeks of gestation without simultaneous urinary infection.From the Medical Birth Registry, we initially identified approximately 1300 cases with clinical information indicating possible pre-eclampsia or eclampsia. After verifying and supplementing this information with details from the hospital records, we identified 307 singleton pregnant women with definite pre-eclampsia. We used a previously described definition of pre-eclampsia in this study . Brieflyn=66) if diastolic blood pressure increased to at least 110 mmHg, along with proteinuria 3+ on dipstick, or at least 500 mg per 24  h. Cases with eclampsia and suspected HELLP  syndrome were regarded interchangeable with severe pre-eclampsia. Cord plasma was analysed for all 66 cases of severe pre-eclampsia.Pre-eclampsia was classified as severe . Single samples were analysed, and procedures suggested by the manufacturer were followed. Oestradiol and AFP were detected in all plasma samples. In both assays, the intra-assay coefficients of variation at high and low levels were always less than 9%.U-test for comparisons between the groups. Differences between proportions were assessed by chi-square tests. The standardised birth weight was divided into four categories: <0.76 corresponds to a strict definition of small for gestational age (SGA), and 0.76–0.89 is a broad category of relatively small infants. The category 0.90 to 1.09 includes infants with appropriate weight for their gestation, and the category >1.09 includes large babies. For each level of birth weight, we estimated values of oestradiol and AFP in the pre-eclampsia group and in controls. In multiple linear regression analyses, we assessed whether the contribution to AFP or oestradiol differed between cases of severe pre-eclampsia and controls, after adjustment for length of gestation, birth weight and offspring sex. All statistical analyses were calculated using the Statistical Package for the Social Sciences (SPSS), version 10.05 .Oestradiol and AFP had positively skewed distributions, and we used Mann–Whitney vs 28.2 years among controls, whereas mean length of gestation was 251 and 280 days. In the pre-eclampsia group, 71% were primipara against 36.4% among controls. Infant birth weight was substantially lower after severe pre-eclampsia (mean: 2527 vs 3565 grams), as was infant length (mean: 45.2 vs 49.9 cm).Table 1P<0.03) in umbilical cord plasma in younger women, but no difference in AFP by maternal age. In primipara, cord plasma oestradiol was higher than in multiparae (P<0.001), and there was a consistent reduction in cord plasma oestradiol (P for trend <0.001) with increasing parity  if the mother reported smoking at the beginning of pregnancy. Birth weight was 106 g higher (P=0.01) in boys than girls. Oestradiol did not significantly differ by sex of the newborn, but AFP was higher in boys than in girls (P<0.02).In Table 2P<0.001). In contrast, there was no difference in oestradiol levels between the groups. Across categories of standardised birth weight (Table 4P=0.01). For oestradiol (Table 6Table 3Severe pre-eclampsia is characterised by abnormally shallow decidual invasion by the trophoblast, placental hypoxia, and reduced uteroplacental blood flow . We founTo our knowledge, no previous study has used umbilical measurements of AFP and oestradiol to examine the association between pre-eclampsia and birth weight. Our study was performed within a population of nearly 13 000 consecutive births  allowed The results in the control population  were in In uncomplicated pregnancies, α-foetoprotein reaches its maximum at the beginning of the third trimester, and drops thereafter towards term (Recent studies have found that women who were born after pre-eclamptic pregnancies have a reduced risk of breast cancer in adult life . In adul


We have examined 645 recorded cases of childhood acute lymphatic leukaemia (ALL) in Sweden during 1973–89 to identify space–time clustering by using the close-pair method of Knox. The records included date of birth and of diagnosis as well as addresses at birth and at diagnosis. There was a significant excess of case pairs close in date of birth and place of birth in the 5- to 15-year age group. © 1999 Cancer Research Campaign