The increased survival of preterm and very low birth weight infants in recent years has been well documented but continued surveillance is required in order to monitor the effects of new therapeutic interventions. Gestation and birth weight specific survival rates most accurately reflect the outcome of perinatal care. Our aims were to determine survival to discharge for a large Canadian cohort of preterm infants admitted to the neonatal intensive care unit (NICU), and to examine the effect of gender on survival and the effect of increasing postnatal age on predicted survival.Outcomes for all 19,507 infants admitted to 17 NICUs throughout Canada between January 1996 and October 1997 were collected prospectively. Babies with congenital anomalies were excluded from the study population. Gestation and birth weight specific survival for all infants with birth weight <1,500 g (n = 3419) or gestation ≤30 weeks (n = 3119) were recorded. Actuarial survival curves were constructed to show changes in expected survival with increasing postnatal age.Survival to discharge at 24 weeks gestation was 54%, compared to 82% at 26 weeks and 95% at 30 weeks. In infants with birth weights 600–699, survival to discharge was 62%, compared to 79% at 700–799 g and 96% at 1,000–1,099 g. In infants born at 24 weeks gestational age, survival was higher in females but there were no significant gender differences above 24 weeks gestation. Actuarial analysis showed that risk of death was highest in the first 5 days. For infants born at 24 weeks gestation, estimated survival probability to 48 hours, 7 days and 4 weeks were 88 %, 70 % and 60 % respectively. For smaller birth weights, female survival probabilities were higher than males for the first 40 days of life.Actuarial analysis provides useful information when counseling parents and highlights the importance of frequently revising the prediction for long term survival particularly after the first few days of life. The improvement in survival rates for preterm and very low birth weight infants has been well documented during the last 20 years -3. More Early reports of survival rates for high risk infants were based on birth weight alone, as assessment of gestational age was relatively imprecise prior to the introduction of routine first trimester ultrasound scans. Subsequently, survival data based on gestational age has become more widely reported ,11. ThisOf further concern is that survival data in many published studies is derived from single tertiary units or collaborations of such centres which are not geographically based ,14. ThisOur aims were to determine survival to discharge for a large Canadian cohort of high-risk infants (<1500 g or ≤30 weeks) representing admissions to 17 NICUs with 75% of tertiary level NICU beds in Canada, and to examine the effects of gestation, birth weight, gender and multiple birth on survival. Utilizing actuarial survival analysis, the effect of increasing postnatal age on expected survival to discharge was investigated.The Canadian Neonatal Network comprised 17 tertiary NICUs across Canada in 1996 . It was Prospective data were collected locally by trained research assistants and transmitted electronically to the Canadian Neonatal Network Coordinating Centre for verification and analysis. Collected data included demographic variables, obstetric information, neonatal illness severity  , therapeGestational age was defined as the best obstetric estimate based on early prenatal ultrasound, obstetric examination and obstetric history, unless the post-natal pediatric estimate of gestation differed from the obstetric estimate by more than two weeks. In that case, the pediatric estimate of gestational age was used instead. An infant was defined as small-for-gestational age (SGA) if the birth weight was less than the 10th percentile for gestational age according to the growth charts established by Arbuckle [SNAP-II [NTISS [Chronic lung disease was defined as oxygen dependency at 36 weeks corrected GA for an infant who was born at ≤32 weeks gestation [Intraventricular hemorrhage (IVH) was defined according to the criteria of Papile [Necrotizing enterocolitis (NEC) was defined according to Bell's criteria (stage 2 or higher) [Retinopathy of prematurity (ROP) was defined according to the International Classification for Retinopathy of Prematurity [Nosocomial infection was defined using blood and cerebrospinal fluid culture results according to Freeman's criteria [Patent ductus arteriosus was defined as clinical diagnosis plus treatment with indomethacin or surgical ligation or both. Seizures were defined as clinically significant episodes witnessed by a nurse or physician and for which anti-convulsant treatment was given. Congenital anomalies were classified according to the WHO International Classification of Diseases, 9 th Revision (ICD-9) [Study variables were defined according to the Canadian Neonatal Network SNAP Project Abstractor Manual . An admiArbuckle  in 1989 [SNAP-II  is a neoI [NTISS  is a scoestation . Intrave higher)  and was maturity  and the maturity . Nosocom (ICD-9) .Data analysis was carried out on two populations – preterm babies with gestation age ≤30 weeks and Very Low Birth Weight (VLBW) babies with birth weight <1,500 grams. A two sample t-test was used to separately test the effect of gender and multiple births on population demographics, adverse outcome and resource use. Similarly a two sample t-test was used to test the effect of gender on survival proportions for each week of gestational age and each 100 grams birth weight range. For the actuarial analysis, the event of interest was death and the time-to-event was the number of days survived. Babies that survived to discharge had their length of stay included as right censored observations. The actuarial survival estimator which estimates the probability of survival to and beyond a set number of days was calculated in daily intervals for the first 60 days of life. The effect of gender, gestation age and birth weight on survival were studied by obtaining actuarial survival estimators for each category. The probability of survival to discharge, for an infant who has survived to a given day in the NICU, was calculated and graphed by gestational age and birth weight category.Between January 8, 1996 and October 31, 1997, 19,507 infants were admitted to participating NICUs. Excluding those with congenital anomalies, there were 3,119 infants ≤30 weeks gestational age and 3,409 infants with birth weight <1,500 grams. Characteristics of the two groups are shown in Table Survival to discharge increased from 45% for infants weighing <600 g at birth to over 95% for those with birth weight >1200 g % higher  were singleton deliveries and 841 (27%) were the products of multiple gestation pregnancies Table . There wS(t); the estimated probability of a baby surviving to at least t days. Babies that were discharged from the NICU can not afford to be ignored as these were most likely to be the longer lived observations. Babies that were discharged from the NICU were therefore included as censored observations in the analysis. By comparing the slope of the curve (which gives the instantaneous risk of death), the highest risk of death is within the first 6 days. Risk of death is higher for smaller babies and babies with lower GA. Risk of death also decreases with time. Figure Actuarial survival curves are given in Figures Since our study is based on a large, geographically defined Canadian cohort, it has greater relevance than those from single units and provides a more realistic picture of neonatal outcomes. In addition, our study cohort was derived from large regional centres with comparable levels of care conforming to current North American standards for neonatal intensive care. Our results are similar to those from previous publications on NICU outcomes, such as that of the NICHD Neonatal Network -4, the VAn advantage in survival for female infants has been reported by previous studies with the greatest difference apparent when stratified by birth weight groups -44. In cActuarial analysis adds a further dimension to standard survival data and highlights several points of interest. For all gestational age and birth weight groups the risk of death is greatest during the first few days of life with relatively few deaths occurring after day 7. This emphasises the importance of revising the chances of survival at regular intervals especially during the first few days of life  declare that they have no competing interests.Huw Jones interpreted data and drafted the manuscript. Stella Karuri performed statistical analysis and interpretation. Shoo K Lee was the principal investigator and interpreted the data and drafted the manuscript. Catherine Cronin, Arne Ohlson and Abraham Peliowski and Anne Synnes were site investigators. All these individuals read and approved the final manuscript. The CNN represents all site investigators, and was responsible for organization and administration of the SNAP study, and subsequent data flow.The pre-publication history for this paper can be accessed here:


Data on the health status of the Roma people in Central and Eastern Europe are sparse and the reasons for their poor health are not clear. The objective of this study was to quantify the differences in birth outcomes between Roma and non-Roma mothers in the Czech Republic and to investigate the potential causes of such differences.A population-based study recruited 8938 non-Roma and 1388 Roma hospitalised singleton births that occurred in two Czech districts (Teplice and Prachatice) between 1995 and 2004. During their stay in hospital, mothers completed a questionnaire on their demographic and socioeconomic characteristics and maternal smoking and alcohol consumption. Data on maternal height and weight and on infants' birth weight and gestational age were taken from hospital records.th percentile of birth weight for gestational age); controlling for all covariates reduced these odds ratios to 1.7 (1.3–2.2), 1.5 (1.1–2.0) and 1.3 (1.0–1.6), respectively. Maternal education made the largest contribution to the ethnic differences; the role of health behaviours was relatively modest.Birth weight and gestational age of Roma infants was 373 (SE 15) g and 0.92 (0.05) weeks, respectively, lower than in non-Roma infants. Controlling for demographic, socioeconomic and behavioural factors reduced these differences to 133 (18) g and 0.57 (0.06) weeks, respectively . In terms of binary outcomes, the Roma vs. non-Roma odds ratios were 4.5 (95% CI 3.7–5.4) for low birth weight (< 2500 g), 2.8 (2.2–3.4) for preterm birth (< 37 weeks of gestation), and 2.9 (2.5–3.4) for intrauterine grown retardation (<10There are striking differences in birth outcomes between Roma and non-Roma mothers. The causes of these differences are complex but largely socioeconomic. The Roma people  are the most important ethnic minority in Central and Eastern Europe. It has been known for a long time that the socioeconomic conditions of most Roma people in Central and Eastern Europe are worse than those of the general population, and it has been suspected that their health follows a similar pattern. However, the main conclusion of two recent reviews was that there is a striking lack of information about the Roma people ,2; even Available data suggest that the health of the Roma people is poor ,2. SimilIn this paper, we analysed a large population based study of births in two Czech communities. The study was originally designed to investigate the effect of air pollution on birth outcomes  but it aAs mentioned above, the study was set up to examine the effects of air pollution . It was During hospitalisation, women completed a self-administered questionnaire with a help of a specially trained nurse. Ethnicity was based on maternal self-report; the dataset contained the following options: Asian; European; Roma, and "other".Mother reported their highest attained education , marital status , smoking (cigarettes per day) and consumption of alcohol (drinks per week) before and during pregnancy. Data on mothers' height (in cm), pre-pregnancy weight (in kg), birth weight (in grams) and gestational age (in weeks) were taken from hospital records. The study attempted to collect data on fathers but we did not use these data because of the high rate of missing values .th gender-specific percentile of birth weight for gestational age) defined as binary variables. Finally, we examined the extent to which the differences between Roma and non-Roma birth outcomes can be explained by available covariates. The statistical models (for both binary and continuous outcomes) were built as follows. First, crude effects of ethnicity were estimated. Second, the ethnic differences were adjusted separately for: (i) demographic characteristics of the infant (gender and district of birth); (ii) maternal demographic characteristics ; (iii) maternal social characteristics ; (iv) maternal size (height and weight); and (v) maternal behaviours . Finally, all these covariates were entered into one model. All analyses were conducted using the STATA statistical software .Because the numbers of Asian and "other" infants were too small for meaningful analyses , the final dataset was restricted to European and Roma babies. The statistical analysis was conducted in several steps. First, socioeconomic and life style variables were tabulated by ethnicity. Second, we compared crude differences between the two ethnic groups in the following birth outcomes: birth weight and gestational age measured as continuous variables, and low birth weight (< 2500 g), preterm birth , and intrauterine growth retardation  occurred among Roma women. The distribution of socioeconomic characteristics and behaviours is shown in table Table Results of logistic regression for binary outcomes were similar table . For lowIn additional analyses, we found no evidence for an interaction between ethnicity and calendar year: the differences between Roma and non-Roma groups remained constant over the study period (not shown in table). There was also no interaction between ethnicity and district; despite the difference in the size of the Roma population between the two districts, the differences between Roma and non-Roma groups were similar in Teplice and in Prachatice (not shown in table).This study in a representative community sample of births found striking ethnic differences in all birth outcomes studied and in most socio-demographic characteristics. The covariates available in the study, particularly maternal education, explained a considerable part of these differences.Several limitations of the study should be considered when interpreting the results. First, more detailed data would provide a more complete picture of the family environment. It is likely that if information on economic and material conditions  would contribute further to the explanation of the ethnic differences. Data on husbands would also help but, as explained above, the proportion of missing data on fathers was too high to be used in the analyses. Data on household assets would also be valuable to improve the assessment of families' socioeconomic status. Similarly, data on factors that could mediate the link between ethnicity, socioeconomic characteristics and birth outcomes, such as nutrition, would be also useful. Finally, ethnic differentials in birth outcomes may be further exacerbated by unmeasured area-level characteristics .Second, some of the measurements might have been inaccurate. In particular, the definition of ethnicity is a difficult issue. Some misclassification is likely; some Roma mothers may have identified themselves as non-Roma. However, self-reported ethnicity, used in this study, has been suggested as the preferred method , and altFinally, the number of Roma births was much smaller than that of non-Roma births, and the unbalanced structure of the sample might have reduced the statistical power. However, given the overall size of the study, the statistical power was sufficient to demonstrate, in Roma vs. non-Roma comparisons, an odds ratio of 1.20 or higher at 95% confidence level. The probability of the beta error is therefore low.In the research on ethnic differences in health in general, an important questions has been debated for some time, namely what are the reasons for such differences . For difMaternal education made by far the largest single contribution to explaining the poor birth outcomes in Roma mothers. This is not surprising, because there were huge differences in educational attainment between Roma and non-Roma mothers in this study table , and becMaternal size, particularly height, partly reflects socioeconomic conditions in earlier life . ReproduThe author(s) declare that they have no competing interests.MB analysed the data and drafted the paper. JD and RJS designed and coordinated the study and commented on the drafts of the paper. IS contributed to the data collection and data management and commented on the drafts of the paper.The pre-publication history for this paper can be accessed here:


The appropriateness of an individual's intra uterine growth is now considered an important determinant of both short and long term outcomes, yet currently used measures have several shortcomings. This study demonstrates a method of assessing appropriateness of intrauterine growth based on the estimation of each individual's optimal newborn dimensions from routinely available perinatal data. Appropriateness of growth can then be inferred from the ratio of the value of the observed dimension to that of the optimal dimension.Fractional polynomial regression models including terms for non-pathological determinants of fetal size  were used to predict birth weight, birth length and head circumference from a population without any major risk factors for sub-optimal intra-uterine growth. This population was selected from a total population of all singleton, Caucasian births in Western Australia 1998–2002. Births were excluded if the pregnancy was exposed to factors known to influence fetal growth pathologically. The values predicted by these models were treated as the optimal values, given infant gender, gestational age, maternal height, parity, and age.The selected sample  comprised 60.5% of the total Caucasian singleton birth cohort. Equations are presented that predict optimal birth weight, birth length and head circumference given gestational duration, fetal gender, maternal height, age and parity. The best fitting models explained 40.5% of variance for birth weight, 32.2% for birth length, and 25.2% for head circumference at birth.Proportion of optimal birth weight (length or head circumference) provides a method of assessing appropriateness of intrauterine growth that is less dependent on the health of the reference population or the quality of their morphometric data than is percentile position on a birth weight distribution. Howevame rate . Currentame rate .This communication seeks to describe, demonstrate and justify the usefulness of an alternative method of assessing the appropriateness of fetal growth from information available in the neonatal record.This method is based on three underlying concepts:1. Appropriateness of growth can be expressed as the ratio of the observed birth dimension to the optimal birth dimension for that an individual neonate. Considering the dimension of weight we refer to this ratio as the proportion of optimal birth weight (POBW): a concept similar to the birth weight ratio . Assessi2. Optimal intrauterine growth is most likely to be achieved during pregnancies unaffected by any maternal or fetal pathology or exposures that can pathologically affect fetal growth,3. The many determinants of fetal growth can be classified as having either a pathological or a non-pathological effect on growth.Factors with non-pathological effects on growth include gestational duration, gender -9, materPathological growth determining factors, such as maternal vascular disease or those associated with congenital malformations, usually restrict fetal growth. More rarely fetal weight is pathologically increased, the most well known example being fetal macrosomia induced by maternal diabetes .pathological or non-pathological those determinants of growth that cannot easily be altered. Multiple pregnancy and maternal race are two examples of such determinants.It is less useful to categorise as either This paper demonstrates our method of assessing the appropriateness of fetal growth by deriving equations for optimal birth weight, birth length and head circumference. Gestational duration, fetal gender and maternal height, age and parity are considered as potential independent variables representing the non pathological determinants of fetal growth. In order to select a population with optimal intrauterine growth, all births with evidence of having been exposed to pathological determinants of fetal growth are excluded. Appropriateness of fetal growth is then expressed as the ratio of the observed birth dimension to the estimated optimal birth dimension for a neonate with the same values for non-pathological determinants of fetal growth. The utility of this approach is discussed.Records of the 126,393 births in Western Australia (WA) during the period of 1998 to 2002 were obtained from the Western Australian Maternal and Child Health Research Database (MCHRDB) . This peOf 1998–2002 WA births, 85% were to Caucasian women and 96.8% were singletons. This example therefore derives standards for singletons born to Caucasian women, see Discussion for generalisability.To achieve a cohort of Caucasian singletons anticipated to exhibit optimal fetal growth, any pregnancy with evidence to suggest that fetal growth may have been affected pathologically must be excluded. Stillbirths and deaths before 28 days were excluded as evidence of a suboptimal intrauterine course, which may be associated with abnormal growth, and, for stillbirths, because duration of intrauterine growth, as opposed to gestational duration to delivery, is not recorded.The selection of further exclusion criteria was guided, in part, by the extensive literature concerning growth restriction as reviewed by Resnik . SuggestSince the primary determinant of birth dimensions is the duration of growth, reliable estimates of gestational duration (GA) are essential, yet exclusion criteria for poor quality gestational data are likely to exclude a biased sample. Details of the algorithm used to obtain the best estimate of GA from all available data are described and justified elsewhere . ApplyinFor each of the three response variables  the Box-Cox transformation  was usedrd, 10th and 90th percentile birth weight were estimated within each completed gestational week for sub-samples of parity and maternal height, to ensure that the model adjusted appropriately for these non-pathological determinants. Finally, to aid clinical interpretation of POBW values, POBW was estimated for the 3rd, 10th and 90th percentile positions by taking the weighted mean of POBWs estimated for each parity/gender stratum of births between 38 and 41 gestational weeks inclusive, assuming a constant maternal height of 162 cm.Maternal height (cm) and maternal age (years) were included as linear predictor variables (see Discussion) centred on the population mean values of 162 cm and 25 years respectively. Infant sex and maternal parity were included as categorical variables. Parity was categorised with first birth as the reference, second and third birth as two separate categories, and fourth and subsequent births constituting the fourth category. Models were fitted using SAS (Version 8.2) . The fit of each model was tested by plotting residuals against GA and against the predicted dependent variable . Additionally, POBWs for 3Table 3 and GA3ln(GA). In multivariate analysis, maternal age was not a significant predictor of birth weight. Parameter estimates for the selected best fitting regression equations for the square root of birth weight are given in Table 2 of 40.5%. The best fitting regression equation for estimating optimal birth weight (grams) can therefore be expressed as:The Box-Cox procedure suggested that square root was the optimal transformation to use for normalising birth weight. The optimal fractional polynomial gestational age terms were GAand, of course,rd, 10th and 90th percentile positions on the birth weight distribution are shown by gestational duration across the range 35–42 weeks in Figure rd, 10th and 90th percentile birth weights are 81%, 87% and 115% respectively, Table This equation suggests that under our standard conditions of birth at 40 weeks gestation to a 162 cm primiparous woman, female infants should weigh 3436.0 g and males 3576.4 g. Second births should weigh 123 g more, third births 158 g more and fourth or subsequent births 189 g more than the first birth. An example of the curves obtained from this equation is shown in Figure 2 and GA3. In multivariate analysis, maternal age was not a significant predictor of birth length. Parameter estimates for the selected best fitting regression equations for the square root of birth weight are given in Table 2 of 32.2%. The best fitting regression equation for estimating optimal birth weight (grams) can therefore be expressed as:The Box-Cox procedure suggested that birth length raised to the power of 0.75 was the optimal transformation to use for normalising birth length. The optimal fractional polynomial gestational age terms were GAand of courserd, 10th and 90th percentile positions of crown heel length were found to be 93%, 95% and 105% respectively, see Table This equation suggests that under our standard conditions females should be 50.3 cm long at birth and males should be 0.83 cm longer. The weighted mean proportions of optimal crown heel length at 32 of 25.2%. The best fitting regression equation for estimating optimal birth weight (grams) can therefore be expressed as:The Box-Cox procedure suggested that it was neither necessary nor desirable to transform head circumference prior to modelling. The optimal fractional polynomial gestational age terms were GA and GAln(GA). All potential predictor variables significantly predicted head circumference, including maternal age. Parameter estimates for the selected best fitting regression equations for the square root of birth weight are given in Table rd, 10th and 90th percentile positions of head circumference were found to be 93%, 96% and 105% respectively, see Table This equation suggest that under our standard conditions and for mothers of 25 years, the optimal head circumference for females was 34.4 cm and for males, 0.61 cm larger. The weighted mean proportions of optimal head circumference at 3These regression equations, derived from a population based sample of more than 62,000 singleton Caucasian pregnancies without the major risk factors for intrauterine growth anomaly, demonstrate the method used at our Institute to assess appropriateness of fetal growth. The method is applicable to all populations with suitable data available. Our results may be directly applicable to other populations besides Western Australian Caucasian and Aboriginal singletons, and we consider it likely to be applicable to all Caucasian populations, but applicability should be verified as suggested below.The use of ratios, such as POBW, in the measurement of intrauterine growth is not a novel idea, ,20 but ha) Ratios, such as POBW, represent continuous interval measures.are in error, significantly altering the estimated value of extreme percentile positions. The positions of percentile extremes are therefore both imprecise and sensitive to actual and perceived data quality. The most precise percentile estimates are those at the highest observation densities, which, since many distributions are akin to Gaussian (particularly those of birth dimensions), is often the 50th percentile or median[b) Estimations of POBW require only a single standard value, the predicted optimal birth weight, rather than values at several points on the birth weight distribution. The precision of estimating a percentile position varies inversely with observation density and, since the majority of distributions have fewer observations at the extremes, extreme observations will be the least precise, whatever the size of the sample generating the distribution. Extreme observations are also most subject to error. When verification of individual observations is not possible (as with de-identified data), it is a common practice to exclude extreme values on the assumption that they th percentile position of the birth weight distribution for a population with a high burden of growth restricting pathologies, but the 8th percentile of a population with optimal fetal growth. For an extreme percentile position to be meaningful therefore, the health status of the population from which it is derived needs to be defined, whereas the predicted birth weight is less sensitive to disease burden. Though less sensitive, the proportion of the reference population with growth disturbing factors will also affect the predicted birth weight, except in the unlikely situation where pathological restriction is balanced by pathological acceleration. For this reason we sought to identify a population without growth disturbing factors. The ratio of observed birth weight to predicted birth weight is more generalisable than extreme percentile positions, and the ratio of observed to predicted optimal birth weight is even more generalisable[c) Births affected by growth disturbing factors are over-represented in the extremes of the growth distribution. Hence the positions of extreme percentiles are sensitive to the incidence of growth affecting pathologies in the reference population and vary with the health of the reference population. For example, a newborn with a POBW of 85% might be at the 20th percentile position) or 3,400 g. In Table d) POBW is a continuous scale that correlates with weight deficit, whereas percentile position is an ordinal scale that does not. For example, Table rd, 10th and 90th percentile positions, of each distribution of weight, length and head circumference at birth. This table of equivalences enables an approximate translation of the literature using percentile positions to percentages of optimal dimensions. The populations from which percentiles are derived will seldom be confined to those without factors affecting growth pathologically. The proportion of optimal equivalences in Table Failure to utilise the advantages of a ratio may in part be due to clinical unfamiliarity. In contrast to percentile position, there is little literature describing the clinical associations of appropriateness of growth expressed as proportions of a desirable birth dimension . For thiThe sample was limited to singleton births to Caucasian mothers. It is not useful to classify all determinants of intrauterine growth according to whether or not they have a pathological effect on growth. For example, twin pregnancy slows fetal growth particularly in the third trimester; and gestation-specific perinatal outcomes for multiple births delivered at term are not as good as those for singletons -24. HoweMaternal race is also a problematic factor. The observed variation in intra-uterine growth rates between ethnic groups ,6 may reThe aim of many of our exclusion criteria was to select a sample of births that had not been exposed to factors that have a pathological effect on intrauterine growth. Although the selection criteria only consider causes of growth anomaly we anticipated that selected births would be more likely to be born at term, be larger and born to taller, older and less disadvantaged mothers for several reasons. For example, pathologically affected growth is more often restricted than accelerated and is also associated with preterm birth and maternal smoking, the most prevalent pathological determinant of intrauterine growth, is associated with maternal age and socio-economic circumstances. Table We believe that the curves shown in Figure (ii) Birth much before term often has a pathological cause that also influences fetal growth. Thus the distributions of birth weights delivered at preterm gestations are no longer Gaussian, but typically, because pathological restriction occurs much more frequently than acceleration, are negatively skewed. Thus weights of neonates born at preterm gestations tend to be lower than fetuses of the same gestational age who go on to deliver at term.In this study we address (i) by using the best estimate of gestational age that can be derived from all available data and to eSome may consider our selection of predictor variables incomplete as it does not include measures of paternal size, maternal weight or maternal weight gain.While paternal size is known to influence fetal growth it was not included because the biological father cannot routinely be identified and therefore measures of paternal size are not available on our database. The proportion of variability accounted for by the regression equations would, no doubt, be increased by the inclusion of paternal height as an independent variable.It has been suggested that maternal size affects fetal growth because it correlates with the area of uterine endometrium available for placentation. Since this area is not directly measurable, it is logical to seek and adjust for the maternal dimension(s) that correlates most closely with it. Maternal height measures skeletal size in the vertical dimension only, while maternal weight is associated with skeletal size and soft tissue mass, including adipose tissue. Skeletal height tends to correlate with skeletal size, but the proportion of weight consisting of soft tissue, particularly adipose tissue, is very variable, weakening the correlation between maternal weight and skeletal size. Data from the 5 month Dutch famine suggest that maternal pre-natal weight for height, a measure of soft tissue mass, is not a strong determinant of birth weight . We therdeterminant of fetal weight, maternal weight gain partially measures fetal growth, whether or not it is optimal and should therefore not be adjusted for when estimating appropriateness of growth.Maternal weight gain is occasionally considered to be a determinant of birth weight . HoweverThe non-pathological determinants of growth used in these models accounted for 40.5%, 32.2% and 25.2% of the variance in birth weight, length and head circumference respectively. The variation between these proportions may result from variation in the accuracy with which each birth dimension can be measured. Birth weight is routinely measured to within 5 g, representing about 0.15% of a median weight baby. Compared with birth weight birth, length is more difficult to measure reliably due to the tendency of the neonate to flex and the facility with which it may be stretched. Measured head circumference at delivery may be influenced by moulding of the head during passage through the birth canal. The effect of moulding on head circumference may be largely avoided by waiting until 2 days after birth before measurement. However with early discharge policies, such a wait risks failing to obtain any measurement of head circumference and in WA head circumference is routinely measured in the delivery room.The highly significant, though small, dependence of head circumference on maternal age, despite adjustment for parity, was unexpected and requires confirmation in independent samplesAll three dependent variables were found to have a linear dependence on maternal height in the range 147–183 cm. Outside this range there was a tendency for regression to the mean value of the dimension. This may occur because we could not include a term for paternal height and there will be a tendency for women at the extremes of height to have partners with heights that are less extreme. However since only 269 (~0.4%) of our selected sample had a height outside this range, maternal height was included as a linear variable.Of the three dependent variables only head circumference had an association with maternal age, which was found to be linear up to age 45 years. Only 16 (0.03%) of our selected sample were older than 45 at the time of delivery, therefore maternal age was also included as a linear variable.In 1963 Lubchenco and colleagues  presenteIn 1993, Wilcox and colleagues  introducThe method reported in this paper introduces two innovations, (i) using optimal, rather than expected, growth as the standard, and (ii) reporting the ratio of the observed to optimal birth dimension as the indicator of appropriateness of growth, rather than a percentile position.We sought a sample with optimal opportunities for fetal growth for the creation of standard both because this is the logical standard and also to avoid the problem of the varying incidence of growth restricting pathology and environmental exposures between populations. Our previously published birth weight standards for Western Australia  excludedThe creation of the standards for optimal fetal growth for Caucasian singletons presented here is possible in part due to additional methods of estimating gestational duration , the abioptimal value for the dimension which we define as the value obtained by regression techniques from a large sample of women without risk factors for intrauterine growth anomaly. In this method, appropriateness of intrauterine growth is expressed as the ratio of the observed birth dimension to the optimal birth dimension rather than as being above or below a specified percentile position of the population distribution of that dimension, avoiding the problems inherent in the use of percentile position. Since POBW is a measure of appropriateness of intrauterine growth it may be used as a continuous variable and subjected to parametric statistical analysis. The use of POBW in clinical and research settings will prove whether it is a more precise predictor of compromise within individuals than previously available indicators of intrauterine growth status.We have presented a comprehensive guide to an alternative method of creating standards for newborn dimensions and assessing appropriateness of intrauterine growth. It is based on the estimation of the GA: best available estimate of gestational age at delivery.POBW: percentage of optimal birth weight.WA: Western Australia.The author(s) declare that they have no competing interests.EB conceived of and directed the study and drafted the manuscript. YL carried out initial statistical analyses. NHdeK gave statistical advice. DML carried out subsequent statistical analyses.All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


The perinatal mortality rate among Indigenous Australians is still double that of the rest of the community. The aim of our study was to estimate the extent to which increased risk of low birthweight and preterm birth among Indigenous babies in Queensland account for their continuing mortality excess. If a large proportion of excess deaths can be explained by the unfavourable birthweight and gestational age distribution of Indigenous babies, then that would suggest that priority should be given to implementing primary health care interventions to reduce the risk of low birthweight and preterm birth . Conversely, if only a small proportion is explained by birthweight and gestational age, then other strategies might need to be considered such as improving access to high-quality hospital care around the time of confinement.Population-based, descriptive study of perinatal mortality rates among Indigenous and non-Indigenous babies, in Queensland, stratified by birthweight and gestational age.Indigenous babies are twice as likely to die as their non-Indigenous counterparts . However, within separate strata of birth weight and gestational age, Indigenous and non-Indigenous rates are similar. The Mantel-Haenszel rate ratio adjusted for birth weight and gestational age was 1.13 . This means that most of the excess mortality in Indigenous babies is largely due to their unfavourable birth weight and gestational-age distributions. If Indigenous babies had the same birth weight and gestational age distribution as their non-Indigenous counterparts, then the relative disparity would be reduced by 87% and 20 fewer Indigenous babies would die in Queensland each year.Our results suggest that Indigenous mothers at high risk of poor outcome (for example those Indigenous mothers in preterm labour) have good access to high quality medical care around the time of confinement. The main reason Indigenous babies have a high risk of death is because they are born too early and too small. Thus, to reduce the relative excess of deaths among Indigenous babies, priority should be given to primary health care initiatives aimed at reducing the prevalence of low birth weight and preterm birth. From a public-health perspective, the major obstetric and perinatal problem in Australia is the poor health of Indigenous mothers and babies. Although Australia as a whole has one of the lowest perinatal mortality rates (PMR) of any of the established market economies , IndigenMoreover, the relative excess of deaths among Indigenous babies has not improved over time. In Queensland in 1987–1989, the PMR was 2.2 times higher among Indigenous compared with non-Indigenous babies and in 2000–2002 the rate was still 2.1 times higher. A similar lack of progress is evident in other states -6.The risk of perinatal death is strongly related to a baby's birthweight and gestational age. For example, the risk of death for a moderately preterm baby (33 to 36 weeks gestation) in Queensland is 8.2 times greater than that of a term baby and for a very preterm baby (27 to 32 weeks) the risk is 41.3 times greater. Similarly, a baby who was born at term but weighed less than 2500 g was 6.3 times more likely to die than a baby who weighed more than 2500 g. In Queensland, Indigenous babies are 1.6  times more likely to be born preterm (<37 weeks gestation) and 2.2  times more likely to be low birthweight at term than non-Indigenous babies .The aim of this paper is to estimate the number and proportion of excess deaths among Indigenous babies that can be explained by their higher risk of low birthweight and preterm birth. This analysis strategy assumes that the distribution of birthweight and gestational age among Indigenous babies reflects the prevalence of antenatal risk factors such as smoking, infection, maternal nutrition and psycho-social stress; while any excess of mortality that remains after the excess low-birthweight and preterm risk among Indigenous babies is removed might say something about the quality of medical care at the time of confinement.Although such reasoning has many proponents -13, riskTo be more specific, if a large proportion of excess deaths can be explained by the unfavourable birthweight and gestational age distribution of Indigenous babies, then that would suggest that priority should be given to implementing primary health care interventions to reduce the risk of low birthweight and preterm birth . Conversely, if only a small proportion is explained by birthweight and gestational age, then other strategies might need to be considered such as improving access to high-quality hospital care around the time of confinement.Data were obtained from the population-based Queensland Perinatal Data Collection for the five years 1998 to 2002. This was the most recent five-year period for which complete data were available. The database includes information on all livebirths and stillbirths of at least 20 weeks gestation or 400 g birthweight. A perinatal death is defined as a stillbirth or the death of a liveborn baby within 28 days of birth. Our data set comprised 231,039 births and 2,255 deaths to non-Indigenous mothers and 13,920 births and 273 deaths to Indigenous mothers. Indigenous status is based on the self-reported Indigenous status of the mother and the gestational age is based on the best clinical estimate, which might be derived from the date of the last menstrual period, ultrasound in early pregnancy or maturity scoring of the neonate at birth. The method used is not recorded.In this paper we report the results of a Mantel-Haenszel procedure, which was used to determine the relationship between perinatal mortality and Indigenous status adjusted for the effect of low birthweight and preterm birth. The proportion and number of deaths that could be avoided were estimated by comparison of the crude and adjusted rates.We also used Poisson-regression models to adjust perinatal mortality rates for birthweight and gestational age. We variously fitted single week or four week categories of gestational age and 250 g and 500 g categories of birth weight. The results were the same as using the Mantel-Haenzel approach with broad categories of gestational age and birthweight, and we included these in preference to the Poisson-regression models for ease of interpretation.2 (3) = 5.78, p = 0.1226) suggesting that the effect of Indigenous status is the same across the birthweight and gestational age strata  and that it is appropriate to use the adjusted combined estimate: M-H Adjusted RR = 1.13, 0.99–1.28.The crude perinatal mortality rate among Indigenous babies was 19.6 per 1000 births, which was 2.01  times higher than the rate among non-Indigenous babies. When perinatal mortality rates were compared within each birthweight and gestational age strata, the point estimates suggested that Indigenous babies were only slightly more likely to die than non-Indigenous babies (Figure These results suggest that if the population of Indigenous babies had the same birthweight and gestational age distribution as non-Indigenous babies, then the relative disparity would be reduced by 87% and there would be 20 fewer deaths of Indigenous babies per year.Perinatal mortality among Indigenous babies has remained twice that of their non-Indigenous counterparts for more than a decade. We found that most of this mortality excess is because Indigenous babies are at greater risk of being born too early and too small. In contrast, the case fatality rates of Indigenous babies who were born preterm or of low birth weight were similar to their non-Indigenous counterparts.Using a framework advocated by several perinatal epidemiologists -13, thesRisk factors for preterm birth and low birth weight include smoking, gentio-urinary tract infections, poor maternal nutrition and psycho-social stress -19. SeveMore specifically, the prevalence of smoking among Indigenous women during pregnancy has been reported to be more than 60%, which is at least 3 times the prevalence for non-Indigenous women ,20,21. AIn Australia, the best example we have of a primary health care initiative aimed at reducing risk factors among Indigenous mothers is the Strong Women Strong Babies Strong Culture program in the Northern Territory . This prAlthough such results are encouraging, if substantial progress is to be made across the whole of Australia, a properly funded national initiative is needed. Such an initiative would include funding to improve access to culturally appropriate primary health care during the antenatal period, which would deliver, inter alia, interventions for smoking cessation, screening and treatment of genito-urinary tract infections, screening for domestic violence, and programs aimed at reducing alcohol consumption and poor nutrition.It would not be a case of one strategy fits all. Instead local partnerships with possibly different types of service models would be needed to implement the national initiative. This approach will encourage creativity, innovation and risk taking, which will be essential ingredients to tackling a situation that has proved difficult to improve.Using vital statistics to set agendas has a long and continuing tradition in public health . The advWith regard to Indigenous status, some mothers may be reluctant to identify as Indigenous, others may be non-Indigenous with an Indigenous male partner, or midwives may not ask the mother or make an educated guess . HoweverFor several reasons, gestational age is known to be less accurate among Indigenous than non-Indigenous babies . NeverthAlthough perinatal mortality rates in Queensland have decreased over the last 16 years, the rates in Indigenous populations remain at least double those in the non-Indigenous population. Our analyses, stratified by birthweight and gestational age, suggest that the priority for reducing the excess mortality among Indigenous babies is primary health care to reduce the prevalence of risk factors during the antenatal period. A primary health care approach encompasses a much-needed component of an overall shift towards empowerment of Indigenous women and increased awareness and ownership of health which has the potential to play an important role in reducing the social inequality that has resulted in outcomes such as those found for perinatal mortality.TJ performed the statistical analysis and participated in drafting the manuscript. MC conceived of the study and its design and participated in drafting the manuscript. Both authors read and approved the final manuscript.


Perfluorooctanesulfonate (PFOS) and perfluorooctanoate (PFOA) are man-made, persistent organic pollutants widely spread throughout the environment and human populations. They have been found to interfere with fetal growth in some animal models, but whether a similar effect is seen in humans is uncertain.We investigated the association between plasma levels of PFOS and PFOA in pregnant women and their infants’ birth weight and length of gestation.We randomly selected 1,400 women and their infants from the Danish National Birth Cohort among those who completed all four computer-assisted telephone interviews, provided the first blood samples between gestational weeks 4 and 14, and who gave birth to a single live-born child without congenital malformation. PFOS and PFOA were measured by high performance liquid chromatography–tandem mass spectrometer.PFOS and PFOA levels in maternal plasma were on average 35.3 and 5.6 ng/mL, respectively. Only PFOA levels were inversely associated with birth weight . Neither maternal PFOS nor PFOA levels were consistently associated with the risk for preterm birth or low birth weight. We observed no adverse effects for maternal PFOS or PFOA levels on small for gestational age.Our nationwide cohort data suggest an inverse association between maternal plasma PFOA levels and birth weight. Because of widespread exposure to these chemicals, our findings may be of potential public health concern. Perfluorooctanesulfonate (PFOS) and perfluorooctanoate (PFOA) are persistent organic pollutants (POPs), which are produced synthetically or from the metabolism of other perfluorinated chemicals (PFCs) . These cPFOA and PFOS are slowly metabolized and have half-lives in humans of around 4 years and 5 years, respectively . They arExposure of rodent dams during pregnancy to both PFOS and PFOA has been associated with decrements in birth weight and length of gestation at birth . AlthougIn this study we examined the relation between maternal plasma PFOS and PFOA levels during pregnancy and birth weight as well as length of gestation using data from the Danish National Birth Cohort .The Danish National Birth Cohort (DNBC) includes data on 91,827 pregnant women from March 1996 to November 2002 . Women wn = 87,752), who provided the first blood sample between gestational weeks 4 and 14 , and who had responded to all four telephone interviews , we randomly selected 1,400 mothers. Of the 1,400 selected, 1,102 had a second blood sample, and 200 were randomly selected among these. Of these 200, 146 had a cord blood sample from the child, and 50 were randomly selected among these. We used the second maternal and cord blood samples to examine whether PFOS and PFOA levels were stable during pregnancy and whether the mother–offspring correlation in blood levels was high. Sample sizes were determined in large part by the capacity of the laboratory to perform PFOS and PFOA assays and by what would be acceptable for the Steering Committee that gives permission for use of cohort data.Among all participants who gave birth to a single live-born child without a reported congenital malformation (http://www.bsmb.dk) (n = 10), we used the expected date of delivery provided by the pregnant women at the second interview (after their ultrasound examinations). Low birth weight (LBW) was defined as birth weight < 2,500 g. Preterm birth was defined as the birth of an infant before 37 completed weeks of gestation (259 days), post-term birth as after 42 completed weeks of gestation (294 days), and term birth as between 37 and 42 completed weeks of gestation. Small for gestational age (SGA) was defined as an infant with birth weight below the 10th percentile at a specific gestational age in weeks, based on all singleton live births of same sex and parity who fulfilled our sampling criteria.Information on potential confounders—such as infant sex, maternal age, parity, socio-occupational status, prepregnancy body mass index (BMI), and smoking during the pregnancy—was collected by highly structured questionnaires (available at bsmb.dk) . Socio-obsmb.dk) . If estiThis research was approved by the UCLA Office for Protection of Research Subjects (Reference No. 06-08-023-01) and the Danish Data Protection Agency (Reference No. J.Nr 2006-41-6324).18O2 PFOS) and PFOA (13C2 PFOA) were used in all extracted procedures. All extractions were performed using solid phase extraction techniques and based on 100 μL of plasma. Details of the analytical methods used are available elsewhere  and PFOA . Only two PFOA values and no PFOS values were reported at the lower limit of quantitation (LLOQ) of 1.0 ng/mL. Values below the LLOQ (n = 2) were assigned a value of half the LLOQ. PFOS and PFOA measurements were determined on the residual plasma component from 15 frozen whole blood samples . These were multiplied by a factor of two to make them comparable to measurements made on plasma  and 95% confidence interval (CI) for low birth weight, SGA, and preterm birth. The PFOS and PFOA levels were entered into the analyses as both continuous and categorical variables . Plasma levels below the 25th percentile were used as the reference group when using categorical variables. Both log-transformed and untransformed blood PFC concentrations were tested in the statistical models, and the findings were similar. For this article, we used only the results of untransformed concentrations. We also conducted regression diagnostics for influential points using Cook’s distance, but no substantial changes in estimates were observed before or after the influential points were excluded, and our results are therefore all based on the whole data set.We adjusted for risk factors that could influence fetal growth or length of gestation at birth. These factors included maternal age, parity, socio-occupational status, prepregnancy BMI, smoking during pregnancy, infant sex, and gestational week at blood drawing. For birth weight, we also adjusted for gestational age at birth, measured in days and entered as linear and quadratic terms into the model to capture the reduced weight gain during the last weeks. All covariates but gestational age at birth and gestational week at blood drawing were introduced into models as indicator variables. Stratified analyses by parity, prepregnancy BMI, term of birth  and sex were also performed to evaluate the consistency of the associations across strata.2. The mean birth weight was 3,623 g, and the incidence of LBW was 1.7%; the mean gestational age at birth was 280 days, and the incidence of preterm birth and post-term birth was 3.8% and 9.9%, respectively , and lowest in the age group of ≥ 35 years (5.1 ng/mL), but after adjustment for parity the difference was reduced to 0.4 ng/mL. High levels were also observed in overweight and obese women .r = 0.62), and after adjustment for PFOS levels in the first maternal blood, the regression coefficient of PFOA became slightly stronger . Analyses stratified by term of birth  revealed that the effects of PFOA on birth weight were more pronounced in preterm and post-term babies  (r = 0.72 for PFOS and 0.84 for PFOA if two outliers were not included)  , but thencluded) , but corTo our knowledge, this is the first nationwide population-based study on birth weight and length of gestation with prospectively collected data on levels of perfluorinated chemicals in maternal sera drawn early in pregnancy. We found only maternal plasma PFOA levels to be inversely associated with birth weight. Risk estimates for preterm birth did not increase monotonically across quartiles of PFOS or PFOA exposure. For all four outcomes evaluated, the outcome in the lowest exposure quartile differed from outcomes in the upper three quartiles, but there was no evidence of a dose–response association above the 25th percentile.Maternal plasma levels in the present study  are similar to most levels reported for U.S. populations, but much lower than those found in fluorochemical workers or residents living near the fluoropolymer production facilities . We founThis study has several strengths. We used state-of-the-art laboratory facilities, rigorous blinding, and hospital records for obtaining data on birth weight and gestational age. We found a high correlation between the concentrations of PFOS and PFOA taken over pregnancy time, but the levels declined over pregnancy time, possibly related to blood volume expansion and decreased serum albumin concentration during pregnancy , changesOur nationwide study has the advantage of being selected from a well-described cohort of > 90,000 pregnant women and their infants . SelectiAdjusting for parity and prepregnancy BMI substantially changed the crude measures of association. For instance, adjustment for only parity in the regression model led to attenuation of the regression coefficient between PFOA and birth weight from −20.5 to −7.2. We did find substantially higher PFOA levels in nulliparous women than in multiparous women, which has been shown before in cord blood , perhapsUnlike the more lipophilic POPs, PFOS and PFOA do not typically accumulate in the lipid tissues because of their oil-repellent properties . BecausePFOS and PFOA can cross the placental barrier, but cord concentrations are usually lower than those observed in maternal serum or plasma . In a reThe association between PFOA and birth weight may or may not have public health importance, even if it is causal. LBW has been associated with higher mortality and morbidity that is not restricted to early years of life , althougThe incidence of LBW and preterm birth in our study population is relatively low due to the selection criteria used for sampling the sub-cohort. This should be taken into consideration when trying to extrapolate our results to other populations. We used clinical recordings of gestational age that were often ultrasound-based. Ultrasound-based estimates are biased if the exposure of interest reduces very early fetal growth. This may explain to some extent the higher incidence of preterm birth we see among the exposed, but analysis of gestational age by exposure category did not provide evidence for such an effect, and such a bias would not explain differences in mean birth weight.We observed an inverse association between PFOA levels and birth weight, although the maternal concentrations were orders of magnitude lower than those at which developmental effects have been observed in the animal toxicity studies. A risk characterization of PFOA shows average sera concentrations associated with the lower 95% confidence limit of the benchmark dose for 10% response levels for various developmental outcomes were at least 29,000 ng/mL in rat . The appIn conclusion, our nationwide cohort study suggests an inverse association between maternal levels of PFOA and birth weight. Although PFOS and PFOA have been phased out by some manufacturers in the United States and in some European countries, they are still produced elsewhere, and other similar compounds are believed to break down to PFOS or PFOA in the environment . If thes


The early-life developmental environment is instrumental in shaping our overall adult health and well-being. Early-life diet and endocrine exposure may independently, or in concert with our genetic constitution, induce a pathophysiological process that amplifies with age and leads to premature morbidity and mortality. Recently, this has become known as ‘programming’ but is akin to ‘maternal effects’ described for many years in the biological sciences and is defined as any influence that acts during critical developmental windows to induce long-term changes in the organisms' phenotype. To date, such delayed maternal effects have largely been characterized in terms of susceptibility to cardiovascular or metabolic disease. Here, we review evidence from experimental animal species, non-human primates and man for an effect of the early-life nutritional environment on adult fecundity and fertility. In addition, using a database of pedigree sheep, we also specifically test the hypothesis that being born small for gestational age with or without post-natal growth acceleration directly programmes fertility. We conclude that there is a lack of compelling evidence to suggest pre-natal undernutrition may directly reduce adult fecundity and fertility, but may exert some effects secondarily via an increased incidence of ‘metabolic syndrome’. Possible effects of being born relatively large on subsequent fecundity and fertility warrant further investigation. We critically review the existing literature, but begin by presenting some new data for a possible effect of early environment (using birth weight as a proxy measure) adjusted for post-natal environment (i.e. growth rate) and controlling for a possible contribution from the shared genomes of mother and offspring. The new data come from five flocks of pedigree Suffolk sheep (n = 2427 ewes) and were analysed using generalized linear mixed model procedures, therefore allowing determination of fixed effects (e.g. fertility) but controlling (blocking) for random effects such as different flocks and shared parentage (i.e. ewe and ram effects). The dataset was kindly provided by Signet Breeding Services .The major non-communicable human diseases of our time are cardiovascular and metabolic, each of which is susceptible to ‘developmental programming’  or to ex2.(a)a). Lambs that were born relatively small for gestational age (SGA) (less than 3 kg), defined as being 2 s.d. below the mean birth weight (5.15 ± 1.1 kg), only exhibited early catch-up growth to weaning at eight weeks of age (b) and not thereafter . As adult ewes, the sheep went on to have between one and 11 litters  lower–upper quartiles) and from 1–19 lambs  lower–upper quartiles). When the ewes were grouped by their birth weight into standard 1 kg categories, and these groups were plotted against their lifetime fertility (derived as average number of lambs per litter), then an interesting and significant  quadratic relationship emerged b, ewe lambs growth restricted at birth had a tendency to regress back to the mean growth rate for that breed during the first eight weeks of life. Since absolute rates of growth will increase linearly with birth weight as a simple mathematical function, we expressed early post-natal growth rate relative to birth weight (as g d−1 kg−1 birth weight) and similarly plotted this value against our value for fertility (i.e. lifetime average number of lambs per litter) in the sheep. Clearly, for this cohort of singleton females, post-natal growth rate did not influence future reproductive success  from the model, providing additional evidence for a lack of any residual genetic basis for our observed effect of birth weight on fertility.However, an important assumption in this dataset, and indeed in any dataset that reports on singleton births, is that each singleton at birth arose following the implantation of a single oocyte. For sheep, at least, we know that this is not always the case. figure 13.(a)et al. et al. A series of studies conducted in Aberdeen, Scotland, during the late 1990s assessed the effects of global maternal nutrient restriction  during pregnancy on aspects of male and female foetal hypothalamic–pituitary–gonadal physiology in sheep per se was evidently unaffected; that is, adult females were still able to become pregnant and have many offspring. Furthermore, litter size greatly improved following rehabilitation, indicating no ‘programmed’, long-lasting effect. Given that, in altricial mammals, the investment in reproductive effort  is 10 times that of humans  diet for 12 successive generations  that the progeny exhibit evidence of delayed reproductive maturity, such as delayed puberty and increased reproductive cycle length (n−6 polyunsaturated fatty acids), by contrast, resulted in an earlier onset of puberty A retrospective, multi-generational (data were available over a 40-year period) analysis was conducted on the breeding success of rhesus monkey females that were born small-for-date (SFD), average-for-date (AFD) or large-for-date and convincingly demonstrated that reproductive outcomes appear to pass down the matrilineal line . That is4.In contrast to the animal studies referred to earlier, there are few studies of note in humans. The Dutch Winter Famine of 1944/45, however, presents a unique database in which to investigate effects on lifetime fecundity . Indeed,in utero growth restriction, whether observed in mice . Therefore, we examined the influence of female-offspring birth weight on her subsequent adiposity and lifetime fecundity. As birth weight increased so did adiposity at 20 weeks of age; for every 1 kg increase in birth weight, the subcutaneous fat depth increased by 0.27 ± 0.01 mm (3.67 (2.67–4.970), median interquartile range) (a). This observation in sheep is also apparent in humans; larger babies tend to become larger adolescents and overweight adults  a. This o5.Taken together, the available evidence suggests that maternal nutrient restriction in mammals may have variable effects on gonadal development in the female foetus and some delayed programmed effects on the basic function of the female offspring's hypothalamic–pituitary–gonadal axis, but reproduction (i.e. lifetime fecundity) appears largely to be a protected function. This is emphasized by the failure to observe significant decrements in fecundity in two species with contrasting reproductive strategies: i.e. rats marginally protein-deprived over many generations . However


Birth weight is negatively associated with cardiovascular diseases and diabetes, but the associations are less well-established in developing populations where birth weight is often unavailable. We studied the association of birth weight and cardiovascular risk, using birth rank as an instrumental variable, in Southern China.We used published data on birth weight by birth rank from an appropriate population and baseline data from the Guangzhou Biobank Cohort Study phases 2 & 3 (2005-8) to examine the adjusted associations, using instrumental variable analysis, of birth weight with clinically measured cardiovascular risk factors and the metabolic syndrome in older (≥ 50 years) men  and women .Estimated birth weight was associated with lower blood pressure , -0.53 to 0.03 and diastolic -0.33 mm Hg 95% CI -0.48 to -0.18 per standard deviation higher birth weight), but had little association with glucose, lipids, waist-hip ratio, body mass index or the metabolic syndrome, adjusted for age, sex, early environment and number of offspring.Birth weight may impact blood pressure; however associations of birth weight with other cardiovascular risk factors may not be related to foetal exposures, but speculatively could be an historical co-incidence, with corresponding implications for prevention. Lower birth weight is associated with higher risk of cardiovascular diseases (CVD) ,2, hyperIn developing populations birth weight, was not until recently, routinely measured and evidence concerning the association of birth weight with adult cardiovascular disease or its risk factors is mainly based on young people  or on smUsing birth rank as an instrumental variable for birth weight provides an alternative approach to observational studies for elucidating the contribution of birth weight to adult cardiovascular disease in general. Such an approach also provides a means of assessing the association in developing populations within a reasonable time frame. Birth rank has been observed to be consistently associated with birth weight in many different settings at various epidemiological stages -30. On tThe Guangzhou Biobank Cohort Study is a collaboration amongst the Guangzhou No. 12 Hospital, the Universities of Hong Kong and Birmingham, and has been described in detail . RecruitThe detailed methods of measurement have been reported . In brieth century from a variety of locations, including Singapore [th century, the male infants were slightly heavier (mean birth weight 3.11 kilograms (kg), standard deviation (SD) 0.39) than the female infants  [The instrumental variable for birth weight was birth rank. We used published equations relating birth rank of order 1 to 10+) to birth weight for 18,425 live born Chinese infants from poor families in Singapore from 1950 to 1951 . This st to 10+ tingapore -27. In aSD 0.37) .We examined eight biological outcome measures, typically associated with cardiovascular disease: systolic blood pressure, diastolic blood pressure, fasting plasma glucose, HDL-cholesterol, total cholesterol, triglycerides, waist-hip ratio (WHR), and body mass index (BMI). We also examined the metabolic syndrome and its components using the International Diabetes Federation definition , with raWe used a separate-sample instrumental variable (SSIV) estimator ,36. SSIVPotential confounders considered were study phase, age (in 5-year age-groups), life course socio-economic position , lifestyle habits , number of offspring and measures of early life experience (leg length and seated height), categorized as in Table Of the 20,086 participants, 18,958 (94.4%) had complete data on all items of interest, so we did not perform imputations. There were more women  than men , and the women were younger (mean age 59.4 (SD 7.1)) than the men (mean age 63.1 (SD 7.0)). Age ranged from 50 to 96 years, but only 457 participants were older than 75 years.Table Table Table th century studies produced similar results  did not vary or foetal exposures impact blood pressure but other mechanisms, of which birth weight is a non-causal marker, underlie the commonly observed inverse association between birth weight and diabetes, such as genetic influences as in the 'foetal insulin hypothesis' . AlternaAlthough, our study is only preliminary and hypothesis generating, nevertheless it has some potential implications. It suggests that the developmental origins of health paradigm needs to be extended beyond birth weight and infancy to investigate how exposures across and within generations affect cardiovascular disease risk factors, and specifically to consider that the observed associations may be a historical co-incidence. It also suggests that there may not be a common foetal development pathway determining blood pressure, diabetes and lipids, consistent with the disparate trends in sub-types of cardiovascular disease with economic development. Moreover, if confirmed it suggests that in developing countries birth weight may not be a useful intervention target.Rather than being determined by birth weight, some of the commonly observed associations between birth weight and cardiovascular diseases may be a marker of exposures across generations or even a historical co-incidence. Future studies of birth weight and cardiovascular risk should include measures of parental and/or grandparental exposures, so that the relative contribution of inter-generational exposures and birth weight can be examined. Moreover, comparisons of maternal exposures during pregnancy across ethnicities and epidemiological stages might help identify the underlying exposures driving associations between birth weight and cardiovascular disease risk. Meanwhile caution should be exercised in extrapolating observational studies from long-term developed populations to other settings.The authors declare that they have no competing interests.CMS originated the idea, designed this analysis, interpreted the results and wrote the first draft, with the help of GML, SLAY and BJC. SLAY revised the analysis and edited the manuscript. THL, KKC, CQJ designed the original study and acquired the data with the help of WSZ. All authors revised the manuscript critically for important intellectual content; and gave final approval of the version to be submitted.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2458/10/270/prepubSupplemental tables. Published data concerning birth rank and birth weight extracted for easy reference from a report from Singapore in 1951/2[in 1951/2 Table S1Click here for fileth century studies[Instrumental variable estimates for the effect per birth weight standard deviation on cardiovascular disease risk factors in 18,958 older Chinese men and women in phases 2 and 3 of the Guangzhou Biobank Cohort Study (2005-8), based on 5 mid 20y studies-27 of biClick here for fileth century studies[Instrumental variable estimates (odds ratio) for the effect per birth weight standard deviation on cardiovascular disease risk factors in 18,958 older Chinese men and women in phases 2 and 3 of the Guangzhou Biobank Cohort Study (2005-8), based on 5 mid 20y studies-27 of biClick here for file


Interpregnancy interval (IPI), marital status, and neighborhood are independently associated with birth outcomes. The joint contribution of these exposures has not been evaluated. We tested for effect modification between IPI and marriage, controlling for neighborhood.short (<12 months), intermediate (12–35 months), or long (36+ months) IPI, or as firstborn. The data had a 2-level hierarchical structure, with births nested in 49 neighborhoods. We used multilevel logistic regression to obtain adjusted effect estimates.We analyzed a cohort of 98,330 live births in Montréal, Canada from 1997–2001 to assess IPI and marital status in relation to small for gestational age (SGA) birth. Births were categorized as subsequent-born with short  and intermediate  IPIs. Subsequent births had a lower likelihood of SGA birth than firstborns. Intermediate IPIs were more protective for married  than unmarried mothers .Marital status modified the association between IPI and SGA birth. Being unmarried relative to married was associated with SGA birth for all IPI categories, particularly for subsequent births with intermediate IPIs is reduced in unmarried mothers. Marital status should be considered in recommending particular IPIs as an intervention to improve birth outcomes.Being unmarried increases the likelihood of SGA birth as the IPI shortens, and the protective effect of  The relationship between interpregnancy interval (IPI) and perinatal health is receiving increasing attention. A recent meta-analysis concluded that the IPI, defined as time between the last delivery and conception of the current pregnancy, can be associated with adverse birth outcomes when the IPI is either too short or too long . The proAlthough research on the IPI has been performed in many countries, relatively few studies have been conducted in developed nations characterized by low rates of adverse birth outcomes and comprehensive health insurance such as Canada . FurtherGiven the above gaps in knowledge concerning the IPI, we sought to determine the relationship between marriage, firstborn birth and subsequent birth categorized according to IPI, and the likelihood of SGA birth, accounting for residential neighborhood cluster variations. We assessed whether effect modification was present between IPI category and marital status, adjusting for neighborhood. The setting was Montréal, a large Canadian city in which SGA birth has been shown to vary according to neighborhood .short IPI (less than 12 months), subsequent-born with intermediate IPI (12 to 35 months), and subsequent-born with long IPI (36 months or more). We did not evaluate very short IPIs because of sample size restrictions. Marital status was defined as being legally married versus not legally married. The following covariates were available for mothers: age , education , country of birth (Canada-born versus foreign-born), and year of birth. The Québec birth registry does not include data on smoking or pregnancy complications. The outcome was defined as SGA birth  versus not SGA [Data were drawn from the live birth registry for the province of Québec, Canada for the years 1997 to 2001. All births to mothers with a residential 6-digit postal code for the city of Montréal were extracted for analysis . The Québec birth registry contains the date of birth of the index and previous birth, but not the conception date. The conception date was calculated by subtracting the gestational age  from the date of birth of the index child. The IPI was then calculated as the months between the conception date and the date of the previous birth. The IPI was used to group births into categories defined as: firstborn, subsequent-born with  not SGA .The complete Canadian 6-digit postal code was used to assign mothers to a police district, the administrative unit used to represent "neighborhood" . Thus thMaternal education was missing for 7,347 births (7.2%). Deterministic imputation was used to replace these missing data using the mean maternal education of the specific postal code area within the police district. The imputation procedure left 23 births with missing maternal education; these values could not be imputed because of the absence of maternal education data in the given postal codes. Maternal country of birth was missing for 1,344 births (1.3%), and the postal code was invalid for 33 births. SGA status could not be determined for 124 births because of implausible gestational age . There wp < 0.0001), we retained this model. The area-level variation corresponded to an intra-class correlation of 0.8%, calculated according to the linear threshold model method [We used multi-level multivariate logistic regression models to estimate adjusted odds ratios, with births  clustered within neighborhoods (specified as the random effect) . Becausel method . The covl method .Individual-level variables were added to the model, followed by neighborhood-level variables. Last, we tested individual-level interactions between IPI category and marital status, and between these variables and other relevant variables. The significance of parameter estimates was assessed using the Wald test. We explored various cut- points for the IPI: 1–12, 13–30, 31+ months; 1–11, 12–23, 24+ months; 1–8, 9–30, 31+ months; and 1–8, 9–35, 36+ months. We chose the <12, 12–35, 36+ month cut-offs based on sufficient sample size in each category, as well as clinical relevance. In addition, we calculated population attributable fractions . AnalyseThis study was conducted through a mandate to monitor and research population health in the province of Québec, Canada, authorized by the Health Ministry and approved by the Québec Public Health Ethics Committee.intermediate for the majority (44.7%), followed by long (37%), and short (18.3%). Among subsequent births, SGA birth was least frequent when the IPI was intermediate (6.6%) compared to short (7.4%) or long  compared to unmarried mothers  and less frequent short (16.7%) or long (33.8%) IPIs, relative to neighborhoods with low perceived security .There was an inverse relationship between unmarried status and neighborhood perceived security Table , and thiMarital status and IPI category were both independently associated with SGA birth. In addition, marital status modified the influence of IPI category on SGA birth. Effect modification was also present between marital status and maternal country of birth. Effect modification was not present between individual and neighborhood variables. Figures i.e., confidence intervals exclude one, Figure short  and intermediate IPIs  than among firstborns . Nevertheless, among foreign-born mothers, a similar pattern of decreasing odds of SGA birth associated with marital status was observed with increasing IPI  for subsequent births with intermediate  and short IPIs .Figure intermediate IPIs were more protective  than those with long IPIs . The likelihood of a SGA birth for subsequent births with short IPIs was not significantly different from those with intermediate IPIs. For unmarried mothers, there was no difference between the likelihood of a SGA birth among the three subsequent birth IPI categories, although the intermediate IPI corresponded to the lowest odds of SGA birth . Firstborns accounted for 26% of the population risk.Our study confirms the results of other studies that have found marital status and IPI to be associated with adverse birth outcomes -16,28-31short IPIs. This association was strongest for Canadian-born mothers. Foreign-born mothers might be less susceptible to health-related consequences associated with being unmarried. We are aware of two previous studies reporting that being unmarried is a greater risk factor for adverse birth outcome in subsequent births compared to firstborns; however, these studies did not address the IPI [Second, through testing for effect modification between individual predictor variables, we demonstrated that the association between IPI and SGA birth depends on maternal marital status. We also showed that the association between marital status and SGA birth varied according to IPI and maternal place of origin (Canadian-born versus foreign-born). Specifically, we found that the likelihood of SGA birth associated with being unmarried was highest for subsequent births compared to firstborns, especially for  the IPI ,28.i.e., short IPI) in a household contributes extra stress to unmarried mothers, thereby negatively impacting the pregnancy environment. In the case of large age gaps between siblings , it might be that that child rearing stresses are diminished and resemble those of unmarried mothers without children. Such a mechanism suggests that a marital partner may be important for diminishing stress associated with caring for younger children. The exact nature of such stressors  remains to be elucidated, however. Also, other unmeasured socio-economic status indicators may partly explain, or confound the observed associations. Maternal age cannot explain the associations because we adjusted for this variable. An alternative interpretation for the influence of marital status is that nutritional depletion may be present in mothers with short IPIs [long IPIs may have had sufficient time to restore nutritional reserves, which may in turn help buffer any adverse effects of being unmarried. Other biological mechanisms may also link the psychosocial stress of being unmarried with the likelihood of SGA birth [Another key finding was that the odds of SGA birth conferred by being unmarried tended to be similar to that of firstborns as the IPI increased. Different mechanisms may be involved. Perhaps the presence of young siblings  associated with lower socioeconomic status. These and other unmeasured risk factors linked to being unmarried may account for or partly mediate the lesser protective effects of IPI among unmarried women.Our study confirmed that firstborns are at greater risk of being SGA than their siblings . Furtherintermediate IPIs were significantly more protective than long IPIs for married mothers only. Thus our results support the recommendation that mothers should avoid prolonged IPIs, but this applies, for unknown reasons, primarily to married mothers. Our data do not support the finding that short IPIs are associated with a greater risk of SGA, and this applies for both married as well as unmarried mothers. We did not evaluate extremely short IPIs in this study.Lastly, our study confirms the association between IPI and SGA birth ,38,39. Oshort or long IPIs (population attributable fraction = 3.2%). One study reported an attributable risk of 9.4% for short or long IPIs, but because the study was restricted to subsequent-born infants (i.e. excluded firstborns) and did not consider marriage, this estimated attributable risk cannot be directly compared with ours [Beyond the influence of IPI on SGA birth, marital status is an especially strong predictor of this outcome. While we suggest neither a causal association nor a strict interpretation of "attributable risk", the estimated attributable fractions indicate that being unmarried (population attributable fraction = 5.3%) is a more important contributor to SGA birth than ith ours .Our study may be subject to several limitations. First, we used broad categorizations of marital status and IPI which may inadvertently mask underlying associations. For example, because our data did not permit finer categorization, we defined "unmarried" as not having a legal marital arrangement; however sub-groups of unmarried women such as those in stable cohabitation may be subject to different associations. Similarly, we categorized foreign-born mothers as one group when in fact differences may exist based on nationality or length of residence in Canada, but this was unavoidable because data on duration of residence is not available in the birth registry. Second, we used an administrative definition of neighborhood that may not correspond to residents' perception of neighborhood; effect estimates might differ for other neighborhood boundaries. Third, we do not have data on potential confounders such as infertility treatment which may partly account for the observed associations, although we excluded multiple births . We do ni.e., intermediate IPI) to mothers contemplating subsequent pregnancies. Our results suggest that married mothers may be more likely to benefit from such recommendations than unmarried mothers. Thus, prevention strategies for unmarried mothers may well need to differ from those for married mothers.Our results have a number of implications for current infant health promotion practices. Present obstetric guidelines focus on promoting an appropriate IPI (Differential benefits to married mothers may be compounded when we consider that the IPI is associated with the social characteristics of the neighborhood Table . AlthougOur results bring into question current public health recommendations in obstetrics that appropriate IPIs should be emphasized as an important intervention for newborn health for all women. Focusing on the IPI as an intervention may be differentially successful depending on the social group a mother belongs to. In fact, unmarried mothers who are most at risk of SGA birth may be the least likely to benefit from such an intervention. Marital status in particular might need to be accounted for in prevention strategies for improving birth outcomes.CI: Confidence interval; IPI: Interpregnancy interval; OR: Odds ratio; SGA: Small-for-gestational-age.The authors declare that they have no competing interests.NA developed the research design, guided the data analysis, interpreted the results, and wrote the manuscript. MD, RWP, and ZCL contributed to the research design, analysis and review of the manuscript. YW performed the statistical analysis. RC contributed to conception pf the study and development of the research design. All authors have seen and approved the final version of the manuscript.The pre-publication history for this paper can be accessed here: