Objective. To compare neonatal morbidity and mortality between late-preterm intrauterine growth-restricted (IUGR) and appropriate-for-gestational-age (AGA) infants of the comparable gestational ages (GAs). Methods. We retrospectively analyzed neonatal morbidity and mortality of 50 singleton pregnancies involving fetuses with IUGR delivered between 34 and 36 6/7 weeks of GA due to maternal and/or fetal indication. The control group consisted of 36 singleton pregnancies with spontaneous preterm delivery at the same GA, in which the infant was AGA. Categorical data were compared between IUGR and AGA pregnancies by X2 analysis and Fisher's exact test. Ordinal measures were compared using the Kruskal-Wallis test. Results. The length of stay of newborns in the nursery, as well as the need for and duration of hospitalization in the neonatal intensive care unit, was longer in the group with IUGR. Transient tachypnea of the newborn or apnea rates did not differ significantly between the IUGR and AGA groups. IUGR infants were found to be at a higher risk of intraventricular hemorrhage. No respiratory distress syndrome, pulmonary hemorrhage or bronchopulmonary dysplasia was observed in either group. The frequency of sepsis, thrombocytopenia and hyperbilirubinemia was similar in the two groups. Hypoglycemia was more frequent in the IUGR group. No neonatal death was observed. Conclusion. Our study showed that late-preterm IUGR infants present a significantly higher risk of neonatal complications when compared to late-preterm AGA infants.  Prematurity is the leading cause of neonatal morbidity and mortality , 2. AnotLate-preterm birth is defined as birth between 34 weeks and 36 6/7 weeks of gestation . During The prevalence of IUGR is high in high-risk pregnancies. As a consequence, this condition is common among elective preterm deliveries and is therefore associated with late prematurity, with the observation of a recent increase in the incidence of these electively delivered late-preterm infants .There are conflicting findings in the literature regarding outcomes of preterm infants with IUGR and appropriate-for-gestational-age (AGA) infants. Neonatal morbidity and mortality have been reported to be decreased –11, unchThe aim of this study was to compare neonatal morbidity and mortality between late-preterm IUGR and AGA infants of the same gestational age.We retrospectively analyzed 86 singleton pregnancies, including 50 pregnancies involving infants with a birth weight of or below the 10th percentile (IUGR) delivered between 34 weeks and 36 6/7 weeks of gestation due to maternal and/or fetal indications. The control group consisted of 36 singleton pregnancies with spontaneous preterm delivery at the same gestational age, in which the birth weight ranged from the 11th to 89th percentile (AGA). The study was performed between 2005 and 2007. Birth weight percentiles were based on the standard growth curve of Alexander et al. . Pregnancies complicated by diabetes  and premature membrane rupture, pregnancies with fetal anomalies and pregnancies with unknown or conflicting dating criteria were excluded. Maternal characteristics included age, preexisting medical problems and pregnancy complications. Delivery characteristics included gestational age at delivery, birth weight, route of delivery, indication of elective birth, and Apgar scores. Neonatal data included death, transient tachypnea of the newborn (TTN), neonatal sepsis, intraventricular hemorrhage (IVH), hypoglycemia, jaundice, total number of days the infant was in the neonatal intensive care unit (NICU), and length of hospital stay. Gestational age at delivery was defined based on the mother's last menstrual period and was confirmed by early ultrasound examination. Preexisting medical problems included hypertensive disorders (chronic hypertension and pre-eclampsia), heart diseases, systemic lupus erythematosus and others . Antepartum complications included oligohydramnios and fetal distress. Possible signs of fetal distress were a constant decrease in fetal heart rate variability, the occurrence of late or variable decelerations upon cardiotocography, or a high systolic/diastolic ratio in the umbilical artery. Amniotic fluid volume was estimated during the evaluation of the fetal biophysical profile. Neonatal acidosis was defined as an arterial umbilical cord pH less than 7.2 .μL (150 × 109/L); (7) apnea of prematurity: prolonged respiratory pause (20 s or longer) accompanied by cyanosis, pallor or bradycardia.Diagnostic criteria for each neonatal problem are applied concurrently by neonatologists as follows: (1) TTN: clinical and radiographic features identified during the first hours of life, followed by characteristic resolution during the subsequent 24–48 hours; (2) neonatal sepsis: positive blood culture and clinical manifestations, or clinical manifestations, radiologic findings and laboratory indicators; (3) IVH: identified by serial cranial ultrasonography ; (4) hypoglycemia: blood glucose level below 40 mg/dL; (5) hyperbilirubinemia; (6) neonatal thrombocytopenia: platelet count less than 150.000/The discharge criteria for preterm infants included weight > 2 kg and good suction upon breast-feeding accompanied by adequate weight gain.X2 analysis and Fisher's exact test. Ordinal measures were compared using the Kruskal-Wallis test. IUGR was considered to be significantly related to outcome when P < .005.Categorical data were compared between IUGR and AGA pregnancies by P > .05). There was a predominance of white women , with no significant difference between groups. Among mothers of the IUGR group, 39 (78%) presented some underlying disease or obstetric complication in addition to IUGR, whereas 11 (22%) did not. Hypertensive syndromes were the most frequent condition and were observed in 24 (48%) women of the IUGR group. Heart disease was observed in 5 (10%) mothers of this group, systemic lupus erythematosus in 4 (8%), and other underlying diseases in 6 (12%)    . Gestational age at delivery ranged from 34 to 36.9 weeks  and did not differ between groups. Preterm induction and preterm cesarean delivery were observed in 39 (78%) women of the IUGR group, whereas 11 (22%) patients went into spontaneous preterm labor. The indications for elective preterm delivery in the 39 patients of the IUGR group included oligohydramnios in 20 cases (51.3%), severe maternal disease in 8 (20.5%), presence of fetal maturity in 2 (5.1%), and abnormalities upon cardiotocography, fetal biophysical profile or umbilical artery Doppler in 9 (23.1%) .P = .0001). The frequency of cesarean sections was 92% in the IUGR group and 25% in the AGA group (P < .0001). No difference in mean umbilical cord pH or the presence of neonatal acidosis was observed between groups. Only one newborn of the IUGR group and none of the AGA infants presented Apgar scores <7 at 5 minutes.There was a significant difference in mean birth weight between the two groups   (ct test) . Howeverct test) .None of the newborns died or developed retinopathy of prematurity or necrotizing enterocolitis during their stay in the nursery.Late-preterm births account for 70% of all preterm births . Since tIUGR is a frequent complication in preterm infants and is the cause of most elective late-preterm deliveries. IUGR may also contribute to the increased morbidity and mortality observed among late-preterm infants. These infants are at risk for hypoglycemia, IVH, prolonged hospital stay and increased need for NICU treatment when compared to AGA infants, thus demonstrating the greater severity of these cases.There is a lack of studies regarding the evolution of late-preterm infants with and without IUGR. Gilbert and Danielsen  reportedAlthough we did not observe more severe neonatal complications in the present cases , hypoglycemia, neonatal sepsis, IVH, thrombocytopenia and hyperbilirubinemia were present in the late-preterm infants studied and were the cause of NICU treatment and prolonged stay in the nursery. The lack of observation of these more severe complications might be explained by the small sample studied since the frequency of these complications is rare in this gestational age group. Mean gestational age was 35.5 weeks in the two groups, a gestational age at which the incidence of respiratory distress syndrome is very low . Thus, tDespite the low rates of severe neonatal complications, these newborns are of marked importance for public health since they account for a large percentage of newborns hospitalized in NICUs and require large amounts of public resources during and after their stay in the nursery –34. In aAs also reported in the study of Gilbert and Danielsen , the freLate-preterm birth poses various risks to the newborn and the obstetrician should always weigh the risks and benefits in each case to decide whether to interrupt pregnancy between 34 weeks and 36 weeks and 6 days of gestation. We believe that the technological advances in obstetrics that have occurred over the last few years permit a better control of high-risk pregnancies. Thus, the priority of the obstetrician is to strive for delivery as close to term as possible.In conclusion, our study showed that late-preterm IUGR infants present a significantly higher risk of neonatal complications and a significantly longer NCIU and hospital stay when compared to late-preterm AGA infants. Thus, evaluation of the birth weight percentile for gestational age may provide a more realistic expectation of outcome among late-preterm infants.


Prenatal life exposures, potentially manifested as altered birth size, may influence the later risk of major chronic diseases through direct biologic effects on disease processes, but also by modifying adult behaviors such as physical activity that may influence later disease risk.We investigated the association between birth weight and leisure time physical activity (LTPA) in 43,482 adolescents and adults from 13 Nordic cohorts. Random effects meta-analyses were performed on categorical estimates from cohort-, age-, sex- and birth weight specific analyses. Birth weight showed a reverse U-shaped association with later LTPA; within the range of normal weight the association was negligible but weights below and above this range were associated with a lower probability of undertaking LTPA. Compared with the reference category (3.26–3.75 kg), the birth weight categories of 1.26–1.75, 1.76–2.25, 2.26–2.75, and 4.76–5.25 kg, had odds ratios of 0.67 , 0.72 , 0.89 , and 0.65 , respectively. The shape and strength of the birth weight-LTPA association was virtually independent of sex, age, gestational age, educational level, concurrent body mass index, and smoking.The association between birth weight and undertaking LTPA is very weak within the normal birth weight range, but both low and high birth weights are associated with a lower probability of undertaking LTPA, which hence may be a mediator between prenatal influences and later disease risk. It has been shown that babies born small have a higher risk of developing type 2 diabetes There is strong evidence that physical inactivity is a major predictor for many chronic diseases and is associated with all-cause mortality The aim of the present study was to explore the nature of the association between birth weight and LTPA later in life in the context of a Nordic collaboration that takes advantage of the use of a standardized meta-analysis method and provided access to data on a large number of adult subjects of both genders and with a wide range of birth weight and ages Datatilsynet).The present study is based on a Nordic longitudinal epidemiologic research program “Prenatal and Childhood Growth in Relation to Cardiovascular Disease” (the NordNet Study). All Nordic cohort studies that could provide information about birth weight and cardiovascular disease were invited to participate in the NordNet Study. The NordNet Study consists of researchers from 13 study centers located in Denmark, Finland, the Faroe Islands, Iceland, Norway and Sweden. They provided access to both published and unpublished raw data from 21 Nordic population-based studies of individuals constituting prospective and/or retrospective cohort studies followed longitudinally over many years and comprising >548.000 subjects in the collective data set. Out of the 13 participating centers, 10 centers Birth weight (in kg) was used as a categorical variable and divided into 11 categories: 0.75–1.25, 1.26–1.75, 1.76–2.25, 2.26–2.75, 2.76–3.25, 3.26–3.75, 3.76–4.25, 4.26–4.75, 4.76–5.25, 5.26–5.75, and 5.76–7.00 kg. The borders of the categories were chosen to minimize the arbitrary effect of digit preference. The birth weight category 3.26–3.75 kg was chosen as the reference category as it contained the most subjects. The birth weight groups 0.75–1.25, 5.26–5.75, and 5.76–7.00 kg were excluded in the analyses as they consisted of few subjects, are special cases, and are more likely subject to errors . Birth wactive or inactive. To construct comparable groups across cohorts, we defined the active category by a similar duration, frequency, and/or intensity of LTPA within age groups with the aim of having similar frequency of inactive subjects across cohorts. The categorizations were created in consultation with the investigators from each cohort.Information on LTPA was obtained by questionnaire including information on duration, frequency, and/or intensity of the LTPA. Based on each cohort's information that consisted of either three to six levels of LTPA or of energy expenditure, we categorized LTPA as 2; kg/m2) was calculated for each subject based on self-reported (3 cohorts) or measured weight and height (10 cohorts) at the time of the physical activity assessment, and smoking habits were recorded on the same occasion.The following potential confounders or mediators were investigated: gestational age (</>37 weeks), duration of education (as defined in each cohort), adult body mass index, and smoking (yes/no). Body mass index , along with the corresponding standard errors, were reported to the coordinating center. Third, the estimated SSRCs were pooled using random-effects meta-analyses techniques with random effects based on strata formed from combinations of cohort-, sex-, and age-categories. Our procedure has been described in detail elsewhere We performed logistic regression of physical activity, modelling the probability of being active, on birth weight used as categorical variable for each combination of cohort and sex and age categories. We also used the SSRCs from the cohort and sex and age specific logistic regression analyses to investigate a potential heterogeneity of the association by age. In these analyses, the subjects were subdivided into two age groups consisting of a comparable number of subjects; a younger group aged 14–34 years and an older group aged 35–74 years. Moreover, we performed meta-regression with age as covariate to test for an age-effect, but there were no significant indications of this being the case. In cohorts where data on potential confounding or mediating factors were available, we performed logistic regressions of LTPA on birth weight used as a categorical variable both adjusted and unadjusted for the potential factors, one at a time, to assess their effects. To assess the impact of self-reported birth weight, we performed a sensitivity analysis omitting the 3 cohorts in which birth weight was self-reported.All cohorts within the context of the NordNet collaboration that were eligible for this study were included irrespective of whether results on birth weight and LTPA had been published or not, and hence, the results of the present study should not be subject to publication bias. Given this analytic framework, our meta-regression analysis was not conducted in conjunction with a traditional systematic review of the literature. There are few published studies on this topic, and those that exist are difficult to compare due to different characteristics of subjects Results from a meta-analysis, pooling all the cohort and age and sex specific regression coefficients from the logistic regression of birth weight as a categorical variable, revealed a reverse U-shaped association between birth weight and LTPA . Lower pAnalyses of birth weight as a categorical variable conducted separately for men and women showed no sex-dependency in the birth weight-LTPA association . The revThe results from the regression analyses of LTPA on birth weight treated as a categorical variable separately for men and women younger than or at least 35 years showed limited evidence of an age modification of the shape or strength of the birth weight-LTPA association .The effect of potential confounders or mediators was investigated separately for men and women. The results are not shown, but are provided in This study investigated birth weight as a determinant of a behavioral outcome in terms of physical activity during leisure time among adolescents and adults. The study indicated that both very low and very high birth weights were associated with a lower probability of undertaking LTPA, whereas the association between birth weight and LTPA was weak in the birth weight interval of 2.76–4.25 kg, which is the range that contains the majority of subjects. These findings were obtained irrespective of adjustment for potential confounders or mediators, although unmeasured or residual confounding cannot be excluded.A few other studies have investigated the association between size at birth and physical activity in adulthood. In 462 middle-aged Finnish men, ponderal index at birth was not associated with the duration of strenuous leisure time physical activity Fewer studies have investigated physical activity in subjects with a high birth weight. The British study included subjects with a high birth weight, and according to the crude prevalence rates of undertaking regular physical activity, those with a birth weight >4.5 kg were less likely to undertake regular physical activity than subjects with a birth weight between 4.0 and 4.5 kg Physical activity is a multidimensional and complex exposure to measure. The studies of size at birth and physical activity have investigated different aspects of physical activity, and this could contribute to the differing results. The studies were conducted in age groups from young adults to the elderly, but we found, however, limited evidence of an age modification of the birth weight-LTPA association.Although the other studies did not explicitly investigate the shape of the association between birth weight and occurrence of LTPA later in life throughout the entire birth weight range, the association seems to be an inverse U-shape and very weak within the normal birth weight range.The underlying mechanisms for the present findings are unknown, but there are several possible explanations. Low birth weight could be associated with a reduced physical capacity through a variety of biological mechanisms, including a reduced muscle strength because of low muscle mass We did not find any obvious confounding or mediating effect of indicators of concurrent social position, body mass index, or smoking status on the birth weight-LTPA association. We cannot, however, exclude the possibility that unmeasured or residual confounding, for example from genes or aspects of social background, may contribute to explain our findings. Gestational age could be regarded as a determinant of birth weight, and adjusting for gestational age strengthened the birth weight-LTPA association indicating that the intrauterine growth pattern, i.e. birth weight for gestational age, may be a stronger determinant for LTPA than the birth weight per se. This finding is in accordance with studies on birth weight and coronary heart disease, which found that the association was strengthened by adjustment for gestational age In the NordNet collaboration, we used a standardized meta-regression method to investigate the association between birth weight and LTPA. Researchers conducting meta-analyses are often limited to published estimates that may not be very consistent with the question of interest. In our study we developed and applied a common analytic strategy. For example, we were able to create a consistent as possible definition of LTPA and to specify which potential confounding variables to adjust for in every cohort, thus allowing us to obtain uniform estimates from each cohort. As with any meta-analysis, there was an assumption that the included cohorts were homogeneous, and these methods helped reduce any impact of non-homogeneity. Additionally, the use of a random effects meta-analysis model allowed any potential heterogeneity in the association between birth weight and LTPA across cohorts to occur around a normally distributed mean effect. The NordNet collaboration circumvented the potential of publication biases usually encountered in meta-analysis on previously published results, as many results that may have had difficulties being published on their own because of null-findings were included in this study. Moreover, this set-up overcame problems experienced in traditional meta-analyses that are unable to adjust for confounding, and allowed for the discovery of the inverse U-shaped birth weight-LTPA association, which would not have been revealed by a simple linear regression.Some information on birth weight and all information on physical activity was self-reported, and thus has an inherent potential for diminished precision. However, a sensitivity analysis showed no influence of self-reported birth weight on the results. There is always a potential for misclassification bias in self-reported estimation of LTPA. However, most likely, any misclassification of LTPA will attenuate the strength of the observed association. Moreover, although simple questionnaires give a relatively crude measure of overall physical activity, they are useful in larger cohort studies This study suggested that factors in early life might influence physical activity behavior in adult life. The results of this study do not necessarily imply recommendations for interventions to modify birth weight; birth weight is a proxy for many determinants, which have not yet fully been identified, and, moreover, birth weight is not easily modifiable In summary, this study provides evidence that both low and high birth weights outside the normal range of birth weights are associated with a lower probability of being physically active during leisure time. This may be part of the underlying mechanism of the association between birth weight and disease and may thus have public health implications.Table S1Logistic regression of leisure time physical activity on birth weight with adjustments among men.(0.01 MB PDF)Click here for additional data file.Table S2Logistic regression of leisure time physical activity on birth weight with adjustments among women.(0.01 MB PDF)Click here for additional data file.


In Africa low birth weight (LBW) (<2500 g), is the strongest determinant of infant morbidity and mortality. The aim of this study was to quantify the effect of maternal anthropometry, education and socio-economic status on gestational age and birth weight.In 1000 Sudanese mothers with singleton births, anthropometric measurements  and newborn birth weight were taken within 24 hours of delivery. Furthermore, maternal education and socio-economic status were recorded. The effect of these maternal variables on gestational age and birth weight was investigated by receiver operating characteristic (ROC) curves and by multivariate logistic regression analysis.Although maternal height was significantly correlated (p = 0.002) with gestational age, we did not find maternal characteristics of value in determining the risk for preterm birth. Birth order was the strongest determinant of birth weight compared to other maternal characteristics. The LBW rate of first born babies of 12.2% was nearly twice that of infants of multiparous mothers. Maternal age and all maternal anthropometric measurements were positively correlated (p < 0.001) with birth weight. A maternal height of <156 cm, a maternal weight of <66 kg, a maternal mid arm circumference of <27 cm and years of education of ≤ 8 years were found to increase the relative risk of LBW but this was statistically significant only in the case of maternal height. Maternal age and BMI had no statistically significant effect on determining the risk for LBW. The social class did not affect the birth weight, while the number of years of education was positively correlated with birth weight (p = 0.01). The LBW rate decreased from 9.2% for ≤ 8 years of education to 6.0% for >12 years of education.Birth order and maternal height were found to be the most important maternal parameters which influences birth weight and the risk for LBW. The duration of maternal education and not social class was found to significantly affect the risk for LBW. There is a large body of literature showing that the world wide problem of low birth weight (LBW), i.e. infants weighing <2500 g, is among the strongest determinants of infant mortality and morbidity. While in industrialized countries the majority of LBW infants do well, thanks to the advances of modern obstetric and neonatal care  the chanBeside biological factors like gestational age (GA), maternal weight and height , life stIn Africa there are much higher percentages of women with low education, poverty and poor nutritional status who are therefore at increased risk of adverse reproductive outcomes including LBW and preterm birth. The identification during pregnancy of such mothers is therefore important in order to determine the level of care and priorities for referral to centres where reasonable obstetric and neonatal care are available. Therefore the aim of this study was to investigate the influence of parity, maternal anthropometry, education and socio-economic status on gestational age and birth weight in a sample of mothers and infants from an inner urban area of Khartoum, Sudan. No such comprehensive study was performed before in Sudan.The study was conducted during a one year period and anthropometric measurements were taken from 1000 mothers and their newborns in the Soba University Hospital in Khartoum, Sudan. The mothers were recruited from a large inner urban area of Khartoum with wide differences in the socio-economic status. The three social classes  were determined by the area of residence. In this study only mothers who were sure of the last date of their menstrual period were included in the study. Mothers who were not sure of their dates or who had multiple pregnancy or had their pregnancy complicated by diabetes mellitus were not included in the study.The study was approved by the Department of Paediatrics of the University of Khartoum and consent was obtained from the mothers.In order to exclude inter-observer variation the measurements were taken within 24 hours of birth by one investigator (EME) in the postnatal wards. Maternal anthropometry included mother weight, height and mid arm circumferences. Mothers weight was measured by a standard scale to the nearest 100 grams. The mothers' height was measured with a standard scale for height to the nearest millimeter and the maternal mid arm circumference was measured by an inelastic tape to the nearest millimeter. Babies' weight was measured by a standard scale  to the nearest 10 grams, The gestational age was calculated from the last menstrual period in completed weeks of gestation.Means and standard deviations (SD) were calculated for all maternal anthropometric parameters, gestational age and birth weight. The relationship between maternal anthropometric parameters, gestational age and birth weight was investigated by correlation analysis. Receiver operating characteristic curves (ROC) were drawn to determine optimal cut-off values of the maternal anthropometric parameters that can point to the relative risk for LBW. The optimal cut-off points are defined by the highest numbers of correct classifications considering the LBW rate. The 95% confidence intervals of the area under the normalized ROC curve (AUC) were calculated as described by Hanley and McNeil . AnalysiMaternal age, anthropometric measurements , number of years of education and social status of 1000 Sudanese women and the characteristics of their newborns are shown in Table With increasing birth order (parity) birth weight increased significantly p < 0.0001) as shown in Table  as shownThe correlation coefficients between maternal quantitative characteristics , gestational age and birth weight are shown in Table In contrast to the gestational age all maternal characteristics were significantly positively correlated p < 0.001) with birth weight. Therefore, ROC curves were drawn and an optimal cut-off point for each parameter was defined as shown in Table  with birUsing the cut-off points shown in Table The effect of social class and education on gestational age and birth weight is shown in Table Additionally to the univariate analysis the effect of birth order, maternal anthropometry and the years of education on preterm birth and LBW was investigated using a multivariate logistic regression model. The evaluation showed that no statistically significant model could be obtained (p = 0.163) describing the influence of maternal characteristics on the probability of preterm birth p (GA<37 weeks).In contrast to the gestational age a statistically significant model (p = 0.0005) for the probability of LBW (p (BW ≤2500 g)) was found:z = 16.90-1.303 · PAR + 0.060 · MA + 0.122 · MW - 0.123 · MH - 0.050 · MMAC - 0.153 · BMI - 0.0340 · YEDWhere PAR is the dichotomized parity , MA is the maternal age in years, MW is the maternal weight in kg, MH is the maternal height in cm, MMAC is the maternal mid arm circumference in cm and YED is number of years of education. However, some parameters in this model were strongly correlated with each other  so that they did not give additional information and were therefore eliminated by the backward selection method. The backward selection of this model showed that the only statistically significant predictors for LBW were parity (p = 0.009) and maternal height (p = 0.006) and the logistic regression model can be simplified to:z = 6.434-0.625 · PAR - 0.054 · MH.This model shows that the probability of LBW is mainly influenced by birth order and maternal height and this result agrees well with the results of the univariate evaluation shown in Fig. The status of the mother nutrition and socio-economic variables have long been known to influence the reproductive performance and outcome and the condition of the infant at birth.In our study using both univariate analysis and confirming by multivariate logistic regression model, we could demonstrate that birth order, maternal anthropometric characteristics and education are of value in estimating the increased risk for LBW. However we could not demonstrate that maternal characteristics can predict the increased risk for preterm delivery as already shown by Voigt et al.  et al inMaternal height was the second most important parameter which influences the risk for LBW in our mothers. Our cut off point of 156 cm agrees well with investigators in Bangladesh  suggesteThe lower predictive value of maternal weight measured at delivery in our study could be due to the high individual differences in the changes that occur in the body weight during pregnancy. In contrast maternal height is not liable to such changes as a result of pregnancy.The duration of maternal education and not maternal social class was found to significantly affect the risk for LBW. Karim et al.  found thOur study confirms the usefulness of anthropometric measurements in identifying mothers at high risk of delivering LBW infants as found by other workers ,21. HoweIn our study birth order was found to be the major factor affecting LBW rate. The influence of the other maternal characteristics is distinctly lower. Nevertheless women with a height of <156 cm and education of ≤ 8 years had a LBW rate of 13.7%. This means that almost one seventh of such mothers will have low birth weight babies. Therefore, we recommend that policy makers should make more emphasis on education as it imparts knowledge and thus modify dietary habits and quality of food consumed. This will lead to a better nutritional status in adolescent girls, resulting in lower rates of LBW and hence great reduction in infant morbidity and mortality.AUC: Area under the ROC curve; CI: Confidence interval: GA: Gestational age; LBW: Low birth weight; ROC: Receiver operatic characteristic; SD: Standard deviation.The authors declare that they have no competing interests.EME and GS had primary responsibility for writing of the manuscript. EME carried out all anthropometric measurements and GS performed statistical analysis. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


Thoracoscopy offers great advantages when compared with open surgery in terms of postoperative pain and pulmonary complications. Considering the benign nature of most of the mediastinal cysts, thoracoscopy is safe and feasible with minimal morbidity. The purpose of this article is to review our experience with four cases of mediastinal cysts resected successfully within a period of one year by thoracoscopy.The cases of mediastinal cysts operated by thoracoscopic excision in K.E.M. Hospital, Mumbai from November 2005 to December 2006 were reviewed. The age varied from six months to 10 years. The patients presented with respiratory distress or recurrent lower respiratory tract infection. All patients underwent Chest X-ray and CT scan thorax to delineate the location of the cyst and its relationship with adjacent vital structures. Two patients had anterior and two had posterior mediastinal cyst. The ports were placed depending on the location of the cyst on the CT scan, following the principles of triangularization. The cysts were excised mainly by blunt dissection.All the patients were successfully managed by thoracoscopic surgery. None of them had intraoperative complications. Dissection in patient with history of recurrent respiratory tract infection was difficult because of adhesions. Intercostal drain was removed within 48hrs and the patients were discharged on the fourth postoperative day.Thoracoscopy in mediastinal cysts is a safe and effective procedure with low morbidity and a shorter hospital stay. Thoracoscopy offers great advantages when compared with open surgery in terms of postoperative pain and pulmonary complications. Thoracoscopy is now being increasingly utilized for various pediatric procedures like patent ductus arteriosus ligation, congenital diaphragmatic hernia repair, tracheo-esophageal fistula repair, lung resection etc. Considering the benign nature of most of the mediastinal cysts, thoracoscopic excision is a safe and feasible procedure with minimal morbidity and therefore finds its application even in asymptomatic patients.Video assisted thoracoscopic surgery permits good exposure of the entire thoracic cavity including the mediastinum and better evaluation of the anatomic relationship. It provides adequate access and space for almost all maneuvers of dissection. Also, the surgical trauma is minimized.The cases of mediastinal cysts operated by thoracoscopic excision in K.E.M. Hospital from November 2005 to December 2006 were reviewed. The age varied from six months to 10 years. The patients presented with complaints of respiratory distress or recurrent lower respiratory tract infection. All patients underwent Chest X-ray and CT scan thorax to delineate the location of the cyst and its association with adjacent vital structures. Magnetic resonance imaging was performed in one patient to rule out intraspinal communication. Two patients had anterior and two had posterior mediastinal cyst.Thoracoscopy was done using conventional endotracheal ventilation. Patient was given lateral decubitus position with padding under opposite hemithorax. The trocars were placed depending on the location of the cyst on the CT scan, following the principle of triangularization. Three 5 mm trocars were used, one for thoracoscope and others for grasper/scissors/hook/aspiration needle. Insufflation of 5 mm Hg was used to create the pneumothorax and to maintain good field of vision. Cyst was first evaluated for its site and relationship with the adjacent structures. The pleura over the cyst was incised and dissected. To make the manipulation easy, cyst was initially aspirated and decompressed. Complete excision of the cyst was done mainly by blunt dissection and traction. Chest tube drain was inserted before the trocars were removed.A five-year-old male child was investigated for persistent breathlessness and cough of two years duration. Patient received treatment for open pulmonary tuberculosis one year back. He also had evidence of rickets with rachitic rosary. Skiagram chest showed persistent haziness in the right upper hemithorax. CT scan  revealedA six-month-old male child presented with breathlessness since seven days. Patient had history of three episodes of lower respiratory tract infection. Skiagram chest revealed a right mediastinal mass with mediastinal shift to the opposite side. CT scan showed a posterior mediastinal cyst with coronal clefting of the seventh thoracic vertebra suggesting a neuroenteric cyst. MRI spine revealed no intraspinal communication. Cyst was initially aspirated . It was A ten-year-old male child presented with recurrent cough. CT scan showed a well-defined cyst in the antero-superior mediastinum suggestive of a pericardial or a thymic cyst. On thoracoscopy, the cyst wall resembled an egg membrane with straw colored fluid within it, diagnostic of a hydatid cyst . To prevA six-year-old male child presented with complaints of fever and chest pain since one week. X-ray and CT scan thorax were suggestive of a cystic lesion in the antero-superior mediastinum. The cyst was separate from the pericardium and was in close relation to the great vessels. On thoracoscopy, there was a large thick walled mediastinal cyst separate from the pericardium, closely adherent to the thymus. Sero-hemorrhagic fluid was aspirated and the cyst was excised. Histopathology was suggestive of a thymic cyst.All the patients were successfully managed by thoracoscopic surgery. None had intra-operative complications. In all patients cyst could be easily dissected from the pericardium, bronchi and major vessels. Size of the cyst varied from 4–6 cms. Dissection in one patient who had history of recurrent respiratory tract infection was difficult. The duration of surgery varied from 80 to 120 mints. Intercostal drain was removed within 48hrs and patients were discharged on the fourth postoperative day. All the patients are asymptomatic over a follow-up period of two months to 12 months.Thoracoscopy in children was first reported by Rodgers and Talbert in 1976.CT scan is considered essential and important for displaying morphology, density and extent of mediastinal cysts. Also it is helpful in delineating any communication with esophagus or bronchus. Although CT scan could not confirm the diagnosis in one of our four cases (case 3), it often accurately defines the relationship of the lesion with the adjacent structures. In cases where a neurenteric cyst is suspected because of associated vertebral anomaly, a careful examination of the spinal extension of the lesion should be done using MRI and the spinal component should usually be approached first.via mediastinoscopy have also been reported as treatment modalities of these cysts. However, the recurrence rates with these methods are much higher than reported following surgical excision.[Percutaneous or transbronchial aspiration, injection of sclerosing agents and excision excision.Thoracoscopy has an expanding role in the treatment of Foregut Duplication cysts. The surgical principle of complete resection is used as followed in open surgery, which helped in reducing the recurrence.et al,[et al,[Michel et al, reportedl,[et al, with thel,[et al, Aspiratil,[et al, The surgIf the cyst shares a common wall with the trachea or esophagus, this portion of the cyst wall is to be left behind and the mucosa is stripped or destroyed by electrocautrey. The abseAdhesions, large cysts and subcarinal cyst are associated with high conversion rates. Compressive cysts with lung distension and mediastinal shift are considered contraindication for thoracoscopy. DissectiTo conclude, thoracoscopy is an effective diagnostic and/or therapeutic method that can and should be included in the operational routine of thoracic surgery for the management of mediastinal cysts. Thoracoscopy is safe with low morbidity and shorter hospital stays and thus reducing the financial burden.


CD34+ and CD34+c-kit+ cells had J-shaped relations and CFU-GM cells had a U-shaped relation with birth weight. Among newborns with ⩾3000 g birth weights, concentrations of these cells increased with birth weight, while those below 3000 g had higher stem cell concentrations than the reference category of 3000–3499 g. Adjustment for cord blood plasma insulin-like growth factor-1 levels weakened the stem and progenitor cell–birth weight associations. The positive associations between birth weight and stem cell measurements for term newborns with a normal-to-high birth weight support the stem cell burden hypothesis of cancer risk.We examined the relation with birth weight and umbilical cord blood concentrations of haematopoietic stem and progenitor populations in 288 singleton infants. Across the whole range of birth weight, there was a positive relation between birth weight and CD34 In a previous study, we demonstrated that the umbilical cord blood concentrations of various haematopoietic stem and progenitor populations correlated with cord blood plasma levels of particular mitogens, especially insulin-like growth factor-1 (IGF-1) (The stem cell burden theory predicts (1) that the + cells, a heterogeneous population of early multipotent stem and progenitor cells, committed progenitors and differentiating cells . Consenting study subjects were recruited from one of two sources: (1) participants in the Worcester, MA-based American Red Cross cord blood program (ACBP), in which a haematopoietic stem cells from umbilical cord blood were collected for possible transplantation, from August 2002 to June 2003, and (2) pregnant women delivering at T-NEMC from October 2004 to April 2006. All the cord blood samples were from full-term  singleton infants. The processing of samples, which includes the determination of cord blood volumes, the determination of initial levels of total nucleated cells (TNC) and mononuclear cells (MNC) before centrifugations or manipulation, the quantitation of haematopoietic stem/progenitor cell populations and the determination of cord blood plasma hormone levels, have been described  CD3ecursors ; (3) CD3 culture ; and (4)Birth weight was first studied as a categorical variable . Geometric means for the stem cell measurements were estimated within the indicated categories of birth weight. Multivariate linear regression was used to examine the association between natural log-transformed measures of stem cell potential (dependent variable) and birth weight , adjusting for maternal and neonatal characteristics . To assess whether there was an underlying linear trend, birth weight was next analyzed as a continuous variable across the whole range of birth weight values with the effect estimates expressed for each 500 g increase in birth weight. The fitted coefficients from the regression analyses were exponentiated to obtain the estimated proportional change in birth weight associated with each independent variable. Statistical significance was set at 0.05 (two-sided). Levels of IGF-1, which had the strongest association with levels of stem cells among the hormones and growth factors examined in a previous analysis . A J-shaped relation was observed for the CD34+ and CD34+c-kit+ cells: for birth weights of 3000 g or greater, stem cell concentrations increased with birth weight, while the lowest category of <3000 g had higher levels than the category of 3000–3499 g. For CFU-GM, an approximate U-shaped relation was observed, with the lowest birth weight category having the highest levels of this cell population . The lowest weight category continued to have the highest CFU-GM cells after adjusting for IGF-1 levels.Adjusting for cord blood plasma levels of IGF-1 in the multivariate analysis weakens the association  can have higher levels of stem/progenitor cell measurements than those with 3000–3499 g birth weight resulting in a J- or U-shaped relation between stem cell levels and birth weight. This finding is intriguing, as J- or U-shaped relationships have often been observed in childhood cancers  and neur+ and CD34+CD38− levels relative to full-term infants   of such infants undergo accelerated growth during the first 6–12 months after birth and attain normal height later in life . Prematu−1, respectively, for the four categories of birth weight in Finally, we examined cord blood plasma IGF-1 levels in relation to stem cell levels and birth weight. Insulin-like growth factor-1 had a positive association with stem cell measurements in our samples (per se, in accordance with the stem cell burden hypothesis.We conclude that there is a J- or U-shaped positive association between birth weight and stem cell measurements, linear among term newborns with normal-to-high birth weight and stronger with the more primitive stem cell sub-population. The nonlinear relationship, however, suggests that birth weight is not an unequivocal indicator of stem cell potential in the context of a prenatal origin of cancer risk. Quantitation of stem cell pools might prove to be a more accurate predictor of cancer risk than birth weight


Low birth weight and high childhood body mass index (BMI) is each associated with an increased risk of coronary heart disease (CHD) in adult life. We studied individual and combined associations of birth weight and childhood BMI with the risk of CHD in adulthood.2 at age seven years was 44% (95% CI: 30% to 59%) compared with individuals with median values of birth weight (3.4 kg) and BMI (15.3 kg/m2).Birth weight and BMI at age seven years were available in 216,771 Danish and Finnish individuals born 1924–1976. Linkage to national registers for hospitalization and causes of death identified 8,805 CHD events during up to 33 years of follow-up (median = 24 years) after age 25 years. Analyses were conducted with Cox regression based on restricted cubic splines. Using median birth weight of 3.4 kg as reference, a non-linear relation between birth weight and CHD was found. It was not significantly different between cohorts, or between men and women, nor was the association altered by childhood BMI. For birth weights below 3.4 kg, the risk of CHD increased linearly and reached 1.28 (95% confidence limits: 1.13 to 1.44) at 2 kg. Above 3.4 kg the association weakened, and from about 4 kg there was virtually no association. BMI at age seven years was strongly positively associated with the risk of CHD and the relation was not altered by birth weight. The excess risk in individuals with a birth weight of 2.5 kg and a BMI of 17.7 kg/mBirth weight and BMI at age seven years appeared independently associated with the risk of CHD in adulthood. From a public health perspective we suggest that particular attention should be paid to children with a birth weight below the average in combination with excess relative weight in childhood. The developmental origins of health and disease (DOHaD) hypothesis proposes an increase in coronary heart disease (CHD) risk in later life with decreasing birth size In addition to non-optimal prenatal growth, postnatal growth 2; kg/m2) during late childhood, and the risk was determined more strongly by the rate of weight gain than the body weight attained Some studies have investigated the combined effects of birth weight and later body weight on CHD risk Eligible cohorts were cohorts participating in a Nordic longitudinal epidemiologic research program “Prenatal and Childhood Growth in Relation to Cardiovascular Disease” (the NordNet Study) Among these cohorts, the research question posed here could be addressed in one large cohort of Danish schoolchildren who ever attended school in the municipality of Copenhagen Information on birth weight in the Finnish cohort was abstracted from hospital birth records, whereas in the Danish cohort the mothers reported it at the first health examination after school entry at approximately seven years of age. Birth weight (in kg) was used as a continuous variable, and the range of birth weight was restricted to 2.0 to 5.5 kg to avoid analyses of outliers due to errors in reporting, recording, or to pregnancy-related and fetal morbidities.2]). BMI at the exact age of seven years was obtained by interpolation between successive measurements around that age or by extrapolation Body height and weight were prospectively measured at school health examinations and recorded in school health records, and were used for calculation of BMI ([weight (kg)]/) indicated by an M in the figure. In the upper left corner of the figure, delimited by a birth weight below 3251 g and a BMI at age seven years above 18.35 kg/m2, the observed HR was in the range from about 1.25 to about 2.0 (th percentile) and a BMI of 17.7 kg/m2 (the 95th percentile) had a risk increase of CHD of 44% (HR = 1.44 [95% CI: 1.30 to 1.59]).Birth weight and BMI at age seven years were weakly correlated (r = 0.18). The distribution of individuals across combinations of intervals of birth weight and BMI at age seven years is shown in bout 2.0 . Howeverbout 2.0 . As an eWe found that the risk of CHD events was inversely related to birth weight among individuals with a birth weight less than 3.4 kg, whereas for individuals with a birth weight above 3.4 kg the association weakened and levelled off from 4 kg and upwards through 5.5 kg. While the association between BMI at seven years and later risk of CHD was strong, highly significant and independent of birth weight as previously reported for the Danish cohort The findings were robust. They were essentially the same for men and women and for both cohorts studied. The cohorts constituted a larger number of individuals followed up for a longer time than any of the previous studies It was not the major objective of this study to address the association between BMI at seven years and risk of CHD in adulthood, which has been carefully investigated and reported on previously on the basis of the large Danish cohort th percentile (2.5 kg) or with a BMI of the 95th percentile (17.7 kg/m2) compared to children with median birth weight and median BMI, respectively (The combined effects of birth weight and weight during childhood on CHD risk have been investigated in some previous studies. One study supports the present finding that the birth weight association with CHD risk is independent of childhood BMI; an analysis of the Aberdeen children of the 1950s cohort showed that adjustment for BMI at school entry did not change the inverse association between birth weight and CHD ectively . Childreectively . SimilarInformation on birth weight in the Finnish cohort and on BMI and CHD in both cohorts was for the present study collected prospectively. Birth weight was in the Danish cohort reported by the parents. This method has been shown to be valid One study showed that the effect of weight at one year on the risk of CHD was not accounted for by parental social class The inverse association between fetal growth and risk of CHD may have several explanations, but two phenomena have received particular attention in the continued investigation of the DOHaD hypothesis Although the change in the association from the inverse relation below 4 kg to apparently no association or, as seen in other studies, even an increase in CHD risk at higher birth weights could not be attributed to a relation to higher childhood BMI, there could still be a mixture of opposite effects. Thus, a possible tendency to further reduction in the risk of CHD by increasing birth weight may have been counterbalanced by high birth weights occurring as a consequence of maternal diabetes or its preceding stages, possibly developed during gestation. This may be associated with later increased offspring risk of diabetes In conclusion, independently of BMI at age seven years, birth weights below 3.4 kg are inversely related to the risk of CHD in adult life, whereas at birth weights above this level the risk weakens, and above 4 kg birth weight is not associated with the risk. Preventing CHD-events by modifying birth weight in the lower range through intervention is not yet possible and may not make sense, because birth weight may be only an indicator of some factors more directly linked to CHD risk. These factors should be searched for and may be modifiable. From a public health perspective we suggest that treatment and prevention of childhood obesity should be given high priority, especially in individuals with a birth weight below the average.


Multivariable analyses showed a positive association between high paternal age (⩾35 years) and acute lymphoblastic leukaemia =0.96–2.31) but no association with maternal age. High birth weight (⩾3500 g) was positively associated with acute lymphoblastic leukaemia . Children of mothers with a previous miscarriage or increased gestation (⩾40 weeks) had reduced risks of ALL . Children born into more crowded households (⩾1 person per room) had substantially lower risks than children born into less crowded homes with also some evidence of a lower risk for children born into homes with three adults . These findings indicate that several early life factors, including living conditions in childhood and maternal miscarriage history, influence risk of acute lymphoblastic leukaemia in childhood.In a historical cohort study of all singleton live births in Northern Ireland from 1971–86  86, 356–361. DOI: 10.1038/sj/bjc/6600012www.bjcancer.comThe Cancer Research Campaign© 2002  The aetiology of childhood leukaemia, the most common childhood malignancy, is poorly understood and there are few established risk factors. Recently there has been growing interest in the relationship between prenatal and early life influences on risk of childhood leukaemia  indicateThere is a lack of consistency among reports examining prenatal and early life associations with childhood leukaemia, suggesting that some of the observed associations are spurious. Study designs utilized to date may have contributed to this lack of consistency and very few population-based cohort studies of childhood leukaemia have been performed ; gestational age (to the nearest complete week); number of previous pregnancies, live births and miscarriages; mother's and father's year of birth; social class (based on father's occupation); method of infant feeding on discharge from place of confinement; number of persons aged above and below 15 years of age per household and; number of living rooms and bedrooms in the household at the time of the birth.n=27) to permit investigation in the same manner.Details of incident cases of leukaemia which occurred before the age of 16 years in Northern Ireland between 1975 and 1997 were obtained from the Northern Ireland Cancer Registry's Register of Childhood Cancers. Data on cases are obtained from multiple sources, including: the Northern Ireland Leukaemia and Lymphoma Register; the Oxford Childhood Cancer Register; clinicians' records and; databases of specific research projects into childhood cancer. Using sex, surname, and date of birth, cases of ALL born in Northern Ireland between 1971 and 1986 were identified within the Child Health System database. Exact matches, and matches where one character was incorrect were accepted after manual checking. Perinatal information relating to the children who subsequently developed ALL was compared to that of children who did not develop ALL. Multiple births were excluded from the analyses. There were too few cases of acute myeloid leukaemia  divided by the number of rooms in the accommodation. These variables and number of previous miscarriages, birth order, diagnosis of Down's syndrome at birth, method of feeding, and number of children and adults in the household were categorized as in Table 1P<0.2 level in univariable analyses were included in multivariable models. As data relating to living conditions were unavailable before 1976 or after 1984, two multivariable models were constructed: model I did not include number of adults in the household or household density; model II included these variables. These models were also adjusted for sex and Down's syndrome and models were also constructed to include only those children with a diagnosis of ALL before the age of 6 years. All statistical analyses were performed using SPSS for Windows, Version 10.0 . Ethical approval was obtained from the Research Ethics Committee of The Queen's University of Belfast.As we did not actively follow up the cohort for migration or death we did not feel justified in calculating person years and modelling the risk of disease using Poisson or Cox regression. Instead we approximated relative risks by the odds ratios obtained from logistic regression models, an approximation which is reasonable for a rare disease such as childhood ALL. Variables with a t-test value 0.1, P=0.92.The Child Health System contained data on 444 168 live births (including 9235 multiple births) which occurred between January 1971 and December 1986 to mothers normally resident in Northern Ireland. The Registrar General's Office, which registers all births within Northern Ireland, registered 444 111 live births for the same period. The Child Health System may include a small number of duplicate records. The Northern Ireland Cancer Registry contains 342 cases of ALL diagnosed in children aged less than 16 years in Northern Ireland between January 1975 and December 1997. Of these, 212 were born between 1971 and 1986, and 189 (89.2%) were identified within the Child Health System. One was a multiple birth and was excluded from the analyses. This study therefore involves 188 cases of childhood ALL and 434 745 non cases. Characteristics of the identified cases are shown in P=0.02), while that associated with high maternal age was 1.33 . Women who had at least one miscarriage were substantially less likely to have a child who developed ALL than women who had no history of miscarriage . Gestation of more than 40 weeks appeared to be protective against future risk of ALL while the risk associated with gestation of less than 38 weeks was similar to that for 38–39 weeks. When gestational age was dichotomized into <40 weeks or ⩾40 weeks, longer gestation had a relative risk of 0.76 . Higher birth weight (⩾3500 g) was associated with a modest increase in risk of ALL . There was no association between risk of ALL and birth order or breast-feeding on discharge from place of confinement. The strongest associations seen were for social class at birth and living conditions in the household into which the child was born. Children whose fathers were in manual occupations at the time of birth had a risk of ALL more than 30% lower than children whose fathers were employed in non-manual occupations. Although number of children in the household at the time of birth was unrelated to risk of ALL, children born into households in which three or more adults were living appeared to have substantially reduced risk of ALL . Furthermore, being born into a more crowded home was associated with a 45% reduction in risk of ALL.P<0.001. The remaining adjusted risks in Model Ia were either unaltered or stronger than the unadjusted risks.Tables 3Further adjustment for number of adults in the household and household density at birth (Model IIa) again strengthened the associations between paternal age, history of miscarriage, and gestational age, and risk of ALL. In this model, paternal age of 35 years or over was associated with an 80% increase in the risk of ALL, while history of miscarriage and longer gestational age were associated with 60 and 50% lower risks, respectively. The negative association between social class and ALL was attenuated while that for birth weight was unaltered. The strong association between household density and risk of childhood ALL seen in the univariable analysis remained in this multivariable model, with children born into less crowded households having almost twice the risk of ALL compared to children from more crowded households.Restricting the multivariate models to only those cases diagnosed before 6 years  than their married counterparts, failure to identify these cases will have reduced the power of the study to detect associations between household density, or young maternal age, and ALL. However, as birth records were identified for almost 90% of possible cases, this is a minor deficiency.We found a reasonably strong positive association between birth weight and risk of childhood ALL. To date, the findings on birth weight and risk of childhood leukaemia have been inconsistent. Several studies have reported positive relationships with ALL . Our stuAlthough higher maternal age has generally been associated with higher risk of ALL (We found gestation beyond 40 weeks to be associated with a substantial decrease in the risk of childhood ALL. Most previous reports failed to find an association between ALL and length of gestation (As in this study, some investigators have shown that higher socio-economic position, is associated with higher risk of ALL (Our finding of around a two-fold greater risk of childhood ALL among children born into surroundings where there is little overcrowding supports the hypothesis that childhood ALL is caused by an inappropriate immunological response to a common infection or infections in children whose immune systems are not ‘programmed’ by early exposure to these infections . In suppThis population based cohort study is the first such study to demonstrate that being born into a household of high socio-economic position or low household density is a risk factor for the development of childhood ALL. There is also evidence that being born into an extended family may be protective. These findings offer support for the theory that modulation of the immune system by early exposure to infection is important in reducing future risk of ALL. Further investigation of this hypothesis is warranted, including the examination of infections that are likely to be transmitted, within households, from adults to children.


The children of teen mothers have been reported to have higher rates of several unfavorable mental health outcomes. Past research suggests several possible mechanisms for an association between religiosity and teen birth rate in communities.The present study compiled publicly accessible data on birth rates, conservative religious beliefs, income, and abortion rates in the U.S., aggregated at the state level. Data on teen birth rates and abortion originated from the Center for Disease Control; on income, from the U.S. Bureau of the Census, and on religious beliefs, from the U.S. Religious Landscape Survey carried out by the Pew Forum on Religion and Public Life. We computed correlations and partial correlations.Increased religiosity in residents of states in the U.S. strongly predicted a higher teen birth rate, with r = 0.73 (p < 0.0005). Religiosity correlated negatively with median household income, with r = -0.66, and income correlated negatively with teen birth rate, with r = -0.63. But the correlation between religiosity and teen birth rate remained highly significant when income was controlled for via partial correlation: the partial correlation between religiosity and teen birth rate, controlling for income, was 0.53 (p < 0.0005). Abortion rate correlated negatively with religiosity, with r = -0.45, p = 0.002. However, the partial correlation between teen birth rate and religiosity remained high and significant when controlling for abortion rate  and when controlling for both abortion rate and income .With data aggregated at the state level, conservative religious beliefs strongly predict U.S. teen birth rates, in a relationship that does not appear to be the result of confounding by income or abortion rates. One possible explanation for this relationship is that teens in more religious communities may be less likely to use contraception. The children of teen mothers in the U.S., on the average, have worse outcomes in a number of ways. They score lower in school achievement tests, have a greater likelihood of repeating a grade, are rated more unfavorably by teachers while in high school, have worse physical health, are more likely to be indicated victims of abuse and neglect, have higher durations of foster care placement, and are almost three times more likely to be incarcerated during adolescence or the early 20 s than the children of mothers who delayed childbearing; the daughters of teen mothers are more likely to become teen mothers themselves.In the United States, what to teach adolescents about sexuality and the prevention of teen pregnancy has been controversial. A number of sex education programs in the U.S. have been mandated to be "abstinence-only" programs, excluding the teaching of contraceptive techniques. As stated in a National Public Radio poll report, "the historical impetus for abstinence education has come from evangelical or born-again Christians.... Eighty-one percent of evangelical or born-again Christians believe it is morally wrong for unmarried adults to engage in sexual intercourse, compared with 33 percent of other Americans....More than twice as many evangelicals as non-evangelicals (49 percent to 21 percent) believe the government should fund abstinence-only programs instead of using the money for more comprehensive sex education [Other polls have presented varying results on similar questions: A 2008 poll in Minnesota  reportedThe connection between religion and attitudes toward contraception prompts investigation of the relationship between religiosity and teen pregnancy.Some studies have suggested that greater religiosity is associated with either greater abstinence or lower teen birth rate. Hardy and Raffaelli, who analyzed data from the National Longitudinal Survey of Youth, reported that higher time one religiosity predicted a lower likelihood of first sexual intercourse between time one and time two . Loury cOther studies have suggested that religiosity is associated with behaviors that could lead to a higher teen birth rate. Studer and Thornton found that among 18 year-olds, religious teenagers were less likely to use medical methods of contraception when sexually active . Dodge aRosenbaum compared adolescents who reported taking a virginity pledge with a matched sample of nonpledgers . Among tSuch a hypothesis is bolstered by the research of Santelli and colleagues, who calculated that 86% of the decline in adolescent pregnancies that occurred between 1995 and 2002 was attributable to improved contraceptive use . SantellA complicating variable related to teen births and religiosity is the rate of abortions among teens. Adamczyk and Felson, after analyzing longitudinal survey data from the U.S., reported that more highly religious women are less likely to have either an abortion or an out of wedlock pregnancy . Tomal, Cahn and Carbone summarized differences in attitudes about family and sexuality between the more religious and conservative U.S. "red families," versus the less religious and more liberal "blue families" . These aThe present study approaches the relationship between teen birth rate and religiosity by looking at data aggregated across states in the United States.This study compiled data from publicly accessible data sets. The data on religiosity were from the U.S. Religious Landscapes Survey, published by the Pew Forum on Religion and Public Life in 2008 . The Pew1) Belief in a God or universal spirit: Absolutely Certain.2) There is only one way to interpret the teachings of my religion.3) Scripture should be taken literally, word for word.4) How important is religion in your life: Very Important.5) My religion is the one true faith leading to eternal life.6) Frequency of attendance at religious services: at least once a week.7) Frequency of prayer: at least once a day.8) How often do you receive a definite answer to a specific prayer request: at least once a month.In the tables published in the Religious Landscapes Survey, the percents reported were aggregated across three pairs of states and across Maryland and the District of Columbia, because the sample size from at least one member those pairs was fairly small. We obtained from the Pew Forum staff the disaggregated data and used those numbers in our data set. The sample sizes were deemed too small in Wyoming and the District of Columbia for Pew Forum to release them, and thus these data points are missing. For Rhode Island, data were missing on two of the eight questions; we imputed these missing data points by means of regression on the remaining six questions, so that Rhode Island could be included in the data set.In order to obtain one composite religiosity score for each state, we averaged the percents of respondents endorsing the most religious answer across the eight questions.The rates of teen birth in the fifty U.S. states plus the District of Columbia were reported by the National Center for Health Statistics at the Centers for Disease Control and Prevention . The datA possible confounding variable in the relationship between teen birthrate and religiosity is household income level. We obtained data on the median household income by state in the U.S. from data published by the U.S. Census Bureau . The medTo account for another factor which could complicate the analysis of teen birthrate and religiosity, we estimated the abortion rate among teenagers for each state. The most recent data available on abortion rates were from 2005, published by the Center for Disease Control . These rThe CDC obtains its abortion rates by surveying the Central Health Departments of the various regions. A different approach is used by the Guttmacher Institute, which surveys providers of abortions. We used the Guttmacher data for 2005 to cross check abortion rates .We examined the intercorrelations among the individual religiosity questions to determine whether these were high enough to form an index score. We then formed an index score by averaging the eight religiosity items.We examined the relationships of the variables with Pearson correlations and partial correlations, as computed by SPSS . The parWe examined the 28 intercorrelations among the eight different religiosity variables reported in the Pew Survey. The minimum intercorrelation was 0.55, and the maximum was 0.96. The average intercorrelation was 0.81. Thus the intercorrelation of the religiosity items are high enough to justify making an index score by averaging the scores across the eight items.For all the correlational analyses reported below, we examined the plots of residuals for the regressions with the same independent variables and with teen birth as the dependent variable. There was a slight trend toward increasing residuals with increasing values of the dependent variable; in our opinion this trend was not enough to invalidate the linear model, in view of the high correlations obtained and the linear appearance of scatter plots.Teen birth rate correlated with our composite religiosity variable with r = 0.73; 95% CI ; n = 49; p < 0.0005. Thus teen birth rate is very highly correlated with religiosity at the state level, with more religious states having a higher rate of teen birth. A scatter plot of teen birth rate as a function of religiosity is presented in Figure 2 = 0.53) whereas these variables with income removed by partialing shared a little over a quarter of their variance (R2 = 0.28).Next we considered whether median family income for states could be a confounding variable. Income negatively correlated with teen birth rate, with r = -0.63, n = 51, p < 0.0005. Furthermore, income correlated negatively with religiosity, with r = -0.66, n = 49; p < 0.0005. Thus the direction and magnitude of correlations made income a primary candidate for a confounding variable. However, the partial correlation of teen birth rate with religion while controlling for income was 0.53; 95% CI ; df = 46; p < 0.0005. Thus the correlation between religion and teen birth remained large and highly significant, even when controlling for income. The raw religiosity scores and teen birth scores shared a little over half their variance . Would including abortion rate as a covariate greatly affect the correlation between teen birth and religiosity? The answer turned out to be no. The partial correlation between teen birth rate and religiosity, controlling for abortion rate, was 0.68; 95% CI ; df = 42; p < 0.0005. The partial correlation between teen birth and religiosity, controlling for both income and abortion rate, was 0.54; 95% CI ; df = 41; p < 0.0005. Thus, even after taking into account the abortion rate and controlling for income, the correlation between religiosity and teen pregnancy remained high and significant.Table When averaging the results of the eight items of the religiosity survey, one approach would be to first compute the z-scores for each item, and then average the z-scores. Such an approach would assign the same standard deviation to each item, so that items with higher standard deviations did not count more heavily toward the average score. When we checked this "average of standardized religiosity scores," its correlation with the average of the raw percents was 0.999 . This result implies that the two measures are interchangeable, and that differences in standard deviations among the eight items do not appreciably influence the distribution of the average religiosity score. We used the raw percents rather than the standardized scores so that the scatterplot would be more intuitively interpretable.Was there something about the averaging process itself that hid important information or inflated the correlation? To check this, we computed the individual correlation with teen birth rate for each of the eight religiosity items. The results are presented in Table Averaging of items probably results in a higher reliability, which would be expected to improve the correlation with teen birth; the higher reliability of the average than of the individual items is predicted by the Spearman-Brown formula . HoweverWe used as the measure of religiosity the percent of respondents replying in the most religious way; how would the conclusion have been affected if we had entered the percent replying in the least religious way? Some of the questions were dichotomies, and thus the correlations for those items would have had the same magnitude with opposite sign. To check a couple of the items that were not dichotomies, we entered the percentages for the least religious response to literal interpretation of scripture and frequency of prayer, and found correlations of -0.91 and -0.90, respectively, between percent most religious and percent least religious. The correlations of teen birth with "irreligiosity" were very similar in absolute value to the correlations with religiosity for those items. The percent who prayed "seldom or never" in the states correlated at -0.67 with teen birth; the percent who felt that scripture was a "book written by men, not the word of God" correlated at -0.63 with teen birth. It appears that alternate scoring mechanisms measuring irreligiosity would yield the same conclusions and would add nothing to our results.The Guttmacher Institute gathers data on abortion rates by contacting providers of abortions rather than central health agencies. We entered into our data set the Guttmacher abortion rate for women 15-44 for 2005; the correlation of Guttmacher abortion rate  with CDC rate is 0.66 . Using the Guttmacher abortion rate rather than the CDC rate in our partial correlations made no substantive changes - for example, the partial correlation of teen birth with religiosity controlling for CDC estimated teen abortion rates was 0.68; the partial correlation controlling for Guttmacher estimated teen abortion rates  was 0.65.The variation among the fraction of teen abortion rates in different states was small enough that it made little difference for the partial correlations whether we used estimated teen abortion rates or the overall abortion rates for women in the state. The partial correlation of teen birth with religiosity, partialling out CDC abortion rates for all women was 0.69; the same partial correlation using overall rates reported by Guttmacher was 0.70. Estimated teen abortions correlated with overall abortions for the state with r = 0.97 for CDC rates and r = 0.99 for Guttmacher rates.To what extent is the main finding reported here, i.e. the correlation between teen birth and religiosity, dependent upon any one state? Inspection of the scatterplot reveals no major outliers; two influential points appear to be those for Mississippi and Utah. Mississippi, as the state both highest in religiosity and teen birth, tends to increase the correlation; however if the correlation is recomputed without Mississippi, the correlation remains in the same region, with r = 0.69. Utah, which is high in religiosity but in the mid range for teen birth, tends to decrease the correlation; with Utah (but not Mississippi) eliminated the correlation between teen birth rate and religiosity would have been 0.76. For Rhode Island, we used imputation to estimate the two out of eight items that were missing; had we simply made Rhode Island a missing data point, the correlation between teen birth and religiosity would have been 0.72. Thus the magnitude of the correlation we report appears not to be greatly altered by the elimination of any one state.At the state level in the U.S., religiosity, as operationally defined by the eight questions of the Pew Survey, accurately predicts a high teen birth rate. The significant and high correlation continues to hold after statistically controlling for income and abortion rate.It is a statistical maxim that higher correlations are to be found using aggregated data, for example state averages, than with individual level data. This is because some of the noise at the individual level is cancelled by the aggregation process, allowing the relationship between signals to be more clear. As stated in an introductory statistics text, "Correlations based on averages are usually too high when applied to individuals." NonethWe should also caution that on an individual level, certain teen pregnancies are often highly desirable, and some teen parents carry out their responsibilities exceptionally well. If it were possible to obtain good data on unplanned teen pregnancy or pregnancy by "immature" teen parents, we would use it, but we did not find such data available. Nonetheless, at the aggregate level, it is probably true that public policies or cultural practices that reduce the overall rate of teen births are, other things equal, desirable.Our findings by themselves, of course, do not permit causal inferences. There could be unstudied confounding variables that account for the correlations we report. But if we may speculate on the most probable explanation, drawing on the other research cited above: we conjecture that conservative religious communities in the U.S. are more successful in discouraging use of contraception among their teen community members than in discouraging sexual intercourse itself.At the level of states in the U.S., conservative religious beliefs predict teen birth rates highly and significantly; the correlation remains high and significant after controlling for income and estimated rates of abortion.The authors declare that they have no competing interests.The study was conceived by JMS, who also retrieved the data sets for the analyses. The authors shared the tasks of data entry, organization of the data, statistical analyses, and preparation of the manuscript. Both approved the final manuscript.