To develop a reference of population-based gestational age-specific birth weight percentiles for contemporary Chinese.Birth weight data was collected by the China National Population-based Birth Defects Surveillance System. A total of 1,105,214 live singleton births aged ≥28 weeks of gestation without birth defects during 2006–2010 were included. The lambda-mu-sigma method was utilized to generate percentiles and curves.Gestational age-specific birth weight percentiles for male and female infants were constructed separately. Significant differences were observed between the current reference and other references developed for Chinese or non-Chinese infants.There have been moderate increases in birth weight percentiles for Chinese infants of both sexes and most gestational ages since 1980s, suggesting the importance of utilizing an updated national reference for both clinical and research purposes. Small for gestational age (SGA) is defined as weighing less than the 10Since gestational age-specific birth weight varies among racial groups th edition) diagnosed for the first time were required to be reported. Surveillance staffs at the community, township, or village levels were responsible for birth information collection, verification, and follow-up. By comparing the data with related data from other systems like Birth Certification, Perinatal Death Registry, etc., the information on reported cases or births are checked for accuracy and completeness. In addition, annual surveys are conducted to identify and correct errors and inaccuracies in the collected data. It is required that the under-reporting rate of live births or malformations should be no more than 1% and errors or missing values on the report form should also be no more than 1%.The NPBDSS was established in 2006, and data collected by the NPBDSS have been included in the official system of the National Bureau of Statistics of China since 2007 The gestational age at delivery was calculated in completed weeks from the first day of the last menstrual period (LMP). In the surveillance areas, women with suspected pregnancy have an ultrasound examination for confirmation according to obstetric clinical guidelines. For women with irregular menses and/or bleeding during pregnancy as well as those who could not remember the LMP, gestational ages were estimated based on their ultrasound examination. Birth weight of each neonate was measured by a trained midwife within one hour after birth, recorded to the nearest 5 g, and included in hospital delivery records. The data were then abstracted by trained surveillance staff and entered into a web-based reporting system st and 99th percentiles. In Alexander et al. From October 2006 through September 2010, a total of 1,153,166 live and still births whose gestation age were equal to or greater than 28 weeks were identified by the NPBDSS. Stillbirth was defined as the delivery of a fetus that has died before birth for which there is no possibility of resuscitation. L (Box-Cox power), M (median) and S (coefficient variation), which are natural cubic splines with knots at each jT (gestation week) as described in Cole and Green's paper th, 50th and 90th percentiles between our data and those from other references as:For statistical analysis, we first conducted a linear regression analysis and investigated maternal and infant characteristics that might affect birth weight. Since fitting smooth curves on sample quantiles of segmented age groups may demand a large sample size and lose information from nearby groups, we utilized the lambda-mu-sigma (LMS) method for the primary analysis of birth weight for specific gestational ages. The LMS method, which has been used in multiple reference curve studies, adopts a Box-Cox transformation based semiparametric technique and solves penalized likelihood equations. The centiles can be briefly summarized by the perc - Chinaperc)/Chinaperc×100. Here, the Chinaperc represents the percentiles calculated from our study, while Otherperc denotes the percentiles published previously.Relative percentual difference  =   of the mothers were Han Chinese, and the rest (6.8%) were minorities. Most (74.0%) mothers aged 20–29 years at the time of delivery, with few (1.3%) aged ≤20 years and 6.7% aged ≥35 years. More than 70% of infants were born to primiparous women (73.0% for boys and 77.0% for girls).  Both matrd, 5th, 10th, 25th and 50th birthweight percentiles for preterm births . In our study, 6.8% of newborns were minorities, similar to the percentage observed in Census 2010 (8.5%). Urban births accounted for 46.4% of our overall study population, and the proportion from Census 2010 was 49.7%. Studies of birth registry data have the potential for error in the estimation of gestational age, measurement of birth weight, and data transcription. To reduce the possibility of error in our study, 0.66% of the records were removed from final analysis due to missing key variables or outlier values. Although variations in birth weight data collected at our various sites could influence the accuracy of percentiles, discrepancies in measurement were likely minimal due to the high quality of prenatal care and professionally trained midwives.A major strength of the current study is the quality of data, which were obtained from the NPBDSS, a large and well-documented national registry designed to represent populations from a large number of geographic locations. Distributions of ethnic and urban-rural groups in the current study are highly comparable to those from the National Census 2010 Click here for additional data file.Table S2Smoothed birth weight percentiles for Han Chinese infants by gender during 2006–2010.(DOCX)Click here for additional data file.


Ninety-four percent of all birth defects and 95% of deaths due to the birth defects occur in low and middle income countries, Tanzania among them. In Tanzania there are currently limited birth defects prevention strategies in place due to limited information on factors associated with the occurrence of birth defects.We conducted a case control study that included newborns born from October, 2011 through February, 2012 at 4 participating hospitals. A case was defined as any newborn of a Dar es salaam resident with a neural tube defect, orofacial clefts, limb reduction defects or musculo-skeletal defects (SBD) born during the study period. A control was defined as the next three newborns (delivered after the case) without birth defects. Univariate, bivariate and multivariate analysis were done using Epi Info version 3.5.1.A total of 400 newborns participated in the study, 100 cases and 300 controls. Factors associated with higher odds of a SBD included maternal fever (adjusted odds ratio (AOR) = 1.99; 95% confidence interval (CI): 1.14-3.52), maternal hypertension , and low birth weight . Antimalarial use during pregnancy was protective . Folic acid supplementation was protective only in bivariate analysis .Maternal fever, hypertension, and low birth weight are associated with higher odds of SBD. Antimalarial use during pregnancy was associated with lower odds of SBD. Early screening of pregnant mothers for hypertension and other causes of low birth weight may reduce SBD in Dar Es Salaam. A birth defect is an abnormality of structure or function which originates during intrauterine development and is evident before birth, at birth or manifests later in life. Structural birth defects (SBD) may be clinically obvious at birth . The mosBirth defects due to teratogens are among the most preventable defects and are more likely to be found in middle and low income countries compared to high income countries , 7. ThisThis study, a retrospective case-control study nested within that birth cohort, included all newborns delivered October, 2011 through February, 2012 in Dar es Salaam at Muhimbili National Hospital (MNH) and the three Municipal hospitals . None of these hospitals are referral hospitals for mothers identified with a birth defects affected fetus. Sample size was calculated using the Fleiss formula with a case control ratio of 1:3; assuming the proportion of controls exposed was 54% , the minTo avoid duplication of interviews when a mother delivered more than one newborn, the following rules were implemented: if all newborns met the case definition, only the first born was included in the study; if only one newborn met the case definition while the other did not, only the case was included in the study; if all newborns met the control definition, only the first born was included in the study. A structured questionnaire was used to interview mothers whose newborns participated in the study. The information included maternal and paternal demographic data and also information on factors that influence structural birth defects. The outcome  was defined as any newborn with diagnosis of a neural tube defect, an orofacial cleft, a musculo-skeletal defect, or any combination of these defects. Bivariate and multivariate analyses were employed, and factors which had a p-value of ≤ 0.2 in the bivariate analysis were included in the multivariate analysis. P-values of less than or equal to 0.05 were considered statistically significant. Newborns with ambiguous genitalia were excluded from analysis. Data were analysed using Epi Info version 3.5.1.A total of 400 newborns participated in the study - 100 cases and 300 controls - with a 97.1% consented to be interviewed.Out of the 400 mothers whose newborns were included, 310 (77.5%) were ages 20 to 35 years with a mean age of 25.9±5.9 years and 361(90.3%) were married. More than half (73.3%) of the 400 mothers had primary education . None ofBivariate and multivariate analyses were employed to assess associations between factors and MSBD . FactorsMaternal hypertension, maternal fever during pregnancy, and low birth weight newborns had higher odds of developing structural birth defects, after accounting for other factors; while antimalarial use during pregnancy was associated with lower odds of structural birth defects. Mothers who reported fever during pregnancy had significantly higher odds s of giving birth to babies affected with congenital anomalies even after adjusting for other factors. Studies conducted in a variety of settings around the world have observed a significant association between maternal fever and birth defects that is consistent with our findings –14. A poThere was an association between newborn birth weight and MSBD. Low birth weight babies (<2500 grams) had higher odds of congenital defects compared to normal birth weight babies (≥2500 grams). This can be potentially explained by several factors. Some birth defects included in our study may increase the risk of polyhydramnios which can result in preterm labour –34. PretIn Dar es sallam, 90.2% of pregnant mothers deliver at a health facility and majority (72%) of them particularly delivered at Muhimbili National Hospital and the three Municipal hospitals respectively , 40. TheThis study identified maternal hypertension, maternal fever during pregnancy, and low birth weight newborns as risks factors associated with major structural birth defects in Dar es Salaam, Tanzania; while antimalarial use during pregnancy was associated with protection against major structural birth defects. By increasing early screening of pregnant mothers for hypertension and other causes of low birth weight or prematurity, through strengthening of the Primary Health Care system where antenatal care visits are done, it may be possible to reduce major structural birth defects in Dar es Salaam, Tanzania. In addition, we recommend promoting utilization of proven reproductive and child health and advocacy strategies to address the identified factors.


Preference for sons in India has resulted in a skewed sex ratio at live birth, probably as a consequence of female feticide. However, it is unclear if these cultural preferences are also currently present in communities who have emigrated from India to England and Wales.Data of all live births in England and Wales from 2007–2011 were obtained from the Office of National Statistics. A logistic regression analysis was used to compare the probability of having a male infant in mothers born inside the United Kingdom (UK) to those born outside the UK, stratified by mothers’ region and country of birth.Mothers born in India were not observed to be giving birth to disproportionately more boys than mothers that were born in the UK , although an excess of male births were observed in mothers born in South-East Asia , the Middle East , and South America . Mothers who were born in Africa were found to be less likely to give birth to boys than girls when compared to mothers born in the UK , and this observation was attributable to women born in East and West Africa.There was no evidence of an excess of males born to women from India in England and Wales. An excess of males were observed in mothers born in South-East Asia, the Middle East and South America. Women born in Africa are less likely to give birth to boys than UK born mothers, an observation that is consistent with previous data. The worldwide sex ratio at live birth  is approximately 1.05  includinWe have used contemporary national statistical data from England and Wales to test the hypothesis that women born in India would be more likely to have male offspring during the period from 2007 to 2011. As this has coincided with a period when many individuals from many countries are arriving in England and having children, we also had the opportunity to assess the probability of having a male infant in women from different regions of the world compared to those born in the United Kingdom.Data was obtained on all live births in England and Wales from 2007–2011 from the Office of National Statistics. Mothers are asked to provide their country of birth when they register the birth of their child (within 42 days of the child being born). The data were provided summarised by region for most individuals except for those who were born in India, Pakistan, Bangladesh and Poland where data were available at the country level.The main outcome variable for this study was the probability of having a male infant compared to a female. The exposure variable was the country or region of the mother’s birth.We calculated the overall sex ratio for each of the areas by dividing the total male births by the total female births. The primary analysis used logistic regression to generate odds ratios for having a male child for mothers born in the geographic area of interest compared to mothers who were born in the UK for all five years of available data (2007–2011). STATA version 12.1 was used to carry out all data analyses . The analysis and manuscript preparation were in accordance with STROBE guidelines.Data were available for 3 552 050 births from 2007 to 2011, increasing from 690 013 births in 2007 to 723 913 births in 2011. Of these, 2 680 239 (75%) were to mothers born in England and Wales and 871 705 (25%) were to mothers who were not born in England and Wales. Data on the maternal country of birth were not available for 106 (<0.01%) births.In the analysis stratified by continent, mothers who were born in Africa were less likely to have a male child than those who were born in the UK, with an odds ratio of 0.98 . Mothers who were born in Europe (excluding UK), the Americas and the Carribean, the Middle East and Asia and Antarctica and Oceania had the same probability of having a male child as mothers born in the UK Table .Table 1Mothers who were born in East Africa and West Africa had lower probabilities of having a male child than those who were born in the UK with odds ratios (OR) of 0.97 (95% CI: 0.96-0.99) and 0.97 (95% CI: 0.96-0.99) respectively Table . Women wMothers who were born in India had no increased probability of having a male child compared to mothers who were born in the UK with an odds ratio of 1.00 (95% CI: 0.98-1.02). Women who were born in Pakistan and Bangladesh had a lower probability of having a male child than women who were born in the UK with odds ratios of 0.98 (95% CI: 0.97-1.00) and 0.98 (95% CI: 0.96-1.00) respectively Table .Table 3Our analysis of national statistical data from England and Wales over a five year period has demonstrated a number of interesting observations. Firstly, there is no evidence that mothers born in India have an excess of male offspring, contrary to our hypothesis. Secondly, mothers from other regions of the world such as the Middle East, South-East Asia and South America have more male infants than those in the reference group of mothers who were born in the United Kingdom. Finally, women who were born in West and East Africa were less likely to have male offspring than the reference group.The strength of this dataset is that it is nationally representative of England and Wales, with data collected on the mother’s country of origin as part of the data collection at the registration of the birth. This is seen in the fact that the mother’s country of birth was missing in less than 0.01% of births registered during the study period. Another strength is that over this time period, England and Wales have experienced relatively high levels of immigration from a wide number of regions of the world. This provides a natural experiment to investigate any potential differences in sex ratio at birth in mothers from differing areas of the world.However, we also need to consider the limitations of this analysis as these are important in interpreting the results. We only have data on the mothers country of birth, and not that of the age of the mother’s arrival in England and Wales, nor the age of the father which obviously is an important factor in sex determination . The refst century ranging from approximately 1.07 to 1.11, while our more recent data from 2007 to 2011 gave a comparable value of 1.053. We used similar data to that which contributed to the prior analysis and did not observe a significant increase in the probability of mothers born in India having male offspring compared to mothers born in the UK. This is consistent with the interpretation that if the practice of sex-selective abortion was present in the last decades of the 20th century, it is less common nowadays. One potential explanation for this change is that the nature of immigrants who leave India for England and Wales has changed [st century are less susceptible to the cultural and economic pressures that create the environment where sex-selective abortion is acceptable. A recent analysis of data from the Norwegian Central Population Register similarly did not demonstrate an excess of male births compared to females from 2006 to 2012 [The sex ratio of males to females born to Indian-born mothers in England and Wales steadily increased from 1969 to 2005, with the most rapid change occurring in the 1990s [Our analysis also identified mothers from countries in South East Asia, the Middle East and South America as having higher odds ratios for having boys than mothers born in the UK. Previous studies have suggested that there has been a rapid increase in the number of males being born in Vietnam , and alsOur data also showed that mothers from East and West Africa had a lower odds ratio of giving birth to boys than mothers from the UK that was unlikely to have occurred by chance. This observation was first reported in 1984 using data from both England and Wales  and the In summary, our data demonstrate no evidence that mothers born in India have an excess of male births when compared to mothers born in the UK, although mothers born in the Middle East, South East Asia and South America do have higher odds ratios for giving birth to boys than mothers from the UK. We also observed that mothers from Africa are giving birth to fewer boys compared to mothers born in the UK. Our data do not allow distinctions between biological explanations and cultural practices, but are hypothesis generating and have the potential to be tested in other countries with relatively high levels of global immigration.No ethical approval was required as fully anonymised data was used; therefore no ethics committee was consulted before proceeding with this project.


Introduction. Universally mothers at 35 years or more have had higher maternal and perinatal risks. This study analyzed the trend of this group in maternal population and determined their risk of having premature children, during the demographic transition period in Chile. Materials and Methods. Epidemiological study conducted in the population of simple live births registered in the Chilean National Database Births of 1991–2012. Analyses were performed in three categories of maternal age: 35 or more, under 35, and 20 to 29 years. The risk of prematurity was measured by crude and Adjusted Odds Ratio from logistic regression model. Results. Mothers aged 35 and older increased in population from 10.6% in 1991 to 16.7% in 2012 and presented an overall prevalence of preterm delivery of 6.7%, higher prevalence than 20–29 age group (4.7%). In aging mothers, the Odds Ratio for preterm birth adjusted for education, marital status, and parity was 1.68 (95% CI (1.66–1.70)) compared to mothers aged 20–29. All differences were significant (p < 0.001). Conclusions. During Chilean demographic transition, mothers aged 35 or older increased steadily and significantly maintaining higher risks of preterm births. Policies to prevent and monitor the late motherhood could contribute to stopping the current trend. In most industrialized countries, the characteristics of the maternal population changed in the last three decades and this had a broad impact on maternal and child health –3.In Chile, the demographic and health statistics indicated changes that were expressed in a decrease in the birth rate and fertility, the latest age at first birth, and other modifications of socioeconomic order such as increased schooling and the female employment rate. These changes resulted from the demographic and epidemiological transition that was occurring since the 80s; in this context, increasing of maternal age corresponded to the demographic structure of population aging , 5.Risks associated with increased late motherhood (from 35 years on) resulted in increased maternal morbidity and mortality. In Chile, advanced maternal age was becoming more difficult to decline in maternal mortality since 2003 where the lowest rate of 12.7/100,000 live births was registered. This was stated by various publications by E. Donoso which showed that the commitment Chile acquired by assuming the 5th Millennium Development Goal by reducing maternal mortality to 75% in 2015  has been impossible to fulfill because of age 4–6 factor and, besides, mortality was concentrated in this group of mothers. Therefore, 51% of maternal deaths that occurred between 2000 and 2009 were associated with comorbidities that mostly affect mothers over 35 years .Along with maternal risk is the high risk of perinatal and neonatal death more specifically , 10 withThe risk was due to the increase in this group of mothers of preterm birth , low birth weight (less than 2500 g) with intrauterine growth restriction , and congenital malformations , 12.Since premature birth rarely responded to a single cause, it considered the presence of other related factors to age, such as chronic diseases, overweight, parity, and socioeconomic status, among others , 14.In Chile, few studies have centered on the impact of advanced maternal age on the births. A previous publication in the same maternal population (1991 to 2008) showed the highest risk for mothers over 38 of having very preterm child  and child with low birth weight .The main objective was to analyze the trend of maternal age population that held newborns unique and alive new born during 1991 to 2011 to determine the risk of having a premature infant with birth among mothers aged 35 or more by comparison to mothers aged 20–29.Epidemiological study of analytic and transversal cut, was made in the population of live births registered in the National Database of Live Births in Chile, during the period 1991–2012. Multiple births and births that do not correspond to the definition of live birth, under 22 weeks gestational age and/or with less than 500 g  were excThe main explanatory variable is the maternal age which was analyzed in three categories: 35 or more, under 35, and 20 to 29 years.The main result was preterm birth defined by World Health Organization, as birth less than 37 weeks gestational age.First, the distribution of three categories of maternal age annually was described and the evolution of prematurity was further evaluated by group of mothers. The trend of prematurity was evaluated with the Prais-Winsten Test. The differences in rates of preterm between 1991 and 2012 were measured by the relative change (% change).The risk of preterm delivery of maternal population aged 35 or older was determined by the crude and Adjusted Odds Ratio (OR) extracted from a model of logistic regression. The adjustment variables were parity, marital status, and educational level of the mother. The reference group was mothers aged 20–29 years, considered as a group of low obstetric risk.Out of a total of 5,459,001 live births in the study period, 95,821 were excluded and 5,363,180 single live births were analyzed.Mothers of 35 or more increased from 10.6% in 1991 to 16.6% in 2012 . Similarp < 0.001). Although it was observed that prematurity significantly increased over time in all groups studied, the group over 34 had a higher initial rate with 6.2% in 1991 and reached 8.0% in 2012 which corresponded to an increase of 29%. This trend was evolving, so it was estimated that prematurity in this group could reach 8.5% in 2015, since the estimated global increase per year was 0.2% by logit function (p < 0.000).In terms of risk measurement, compared to mothers aged 20–29 years, the Adjusted Odds Ratio (AOR) for parity, marital status, and education level, for preterm birth in women aged 35 years and over, was 1.68 (95% CI (1.66 to 1.70)).The risk increases with age and primiparity and it was observed that there was a significant increase in primiparity in mothers aged 35 and over, from 11.3% to 16.0% at the end of the observed period. In this specific group of primiparae, the risk of having a premature child increased significantly: AOR = 2.10 (95% CI (1.93 to 2.27)). This trend became more evident since 2009 .In summary, in Chile, over the past two decades, premature births increased overall and in all ages. The population groups of mothers of 35 or older and also primiparas increased and this group had higher rates of preterm delivery over time.The adjusted OR for level of education and marital status remained unchanged, which meant that elderly maternal age is a premature birth risk factor independent of schooling and marital status but it was not independent from parity, because primiparity significantly increased the risk of mothers of 35 years.The present study provided information to evaluate emerging problem of maternal and perinatal health, a public health problem as part of changes in population in demographic transition period . Indeed,The increase in mothers of 35 or more deserved special attention, mainly in countries with limited resources or those at low income, since the new situation leads to multiple challenges to obstetric and perinatal care such as increasing human teams and highly specialized technicians for both the mother and the newborn. It was therefore necessary to focus prevention strategies towards women of childbearing age, as appropriate and relevant information about maternal and perinatal risk associated to age, older age at first birth.Large databases allow seeing the demographic reality, a reality on a large scale, and therefore constitute strong evidence. However, they have a limitation which is not allowing access to more specific risk profiles, especially biological ones. Otherwise, it has been known by national and international publications that hypertensive and metabolic pregnancy disorders, especially diabetes mellitus and obesity, are the main risk factors added to advanced maternal age –21.Studies from industrialized countries indicated not only these diseases as the most prevalent and described similar results but also risks associated with treatments for hypofertility used in elderly women that led to a significant increase in multiple pregnancies.Moreover, in developed countries, increase of maternal age and late primiparity are related to the more important demand of assisted reproductive technologies (ARTs). In Chile, medical offer of assisted procreation has increased since the decade of 90s in major cities. This offer remains with a high individual cost; however, it has a growing acceptance by the population .Therefore, the contribution of this group to national situation of preterm birth in the recent twenty years seems to be relatively poor. Indeed, those births have not reached an average of 500 per year and 26% of them were born prematurely . These pThe results presented could be useful in other Latin American countries that shared similar realities with Chile and started the process of demographic transition in the case of Bolivia, Ecuador, Peru, and Venezuela just to mention a few .Future researches should investigate the factors influencing the moment of motherhood in the population, usually noncollected information, in order to target prevention efforts.


Nigeria continues to have one of the highest rates of neonatal deaths in Africa. This study aimed to identify risk factors associated with neonatal death in Nigeria using the 2008 Nigeria Demographic and Health Survey (NDHS).Neonatal deaths of all singleton live-born infants between 2003 and 2008 were extracted from the 2008 NDHS. The 2008 NDHS was a multi-stage cluster sample survey of 36,298 households. Of these households, survival information of 27,147 singleton live-borns was obtained, including 996 cases of neonatal mortality. The risk of death was adjusted for confounders relating to individual, household, and community level factors using Cox regression.Multivariable analyses indicated that a higher birth order of newborns with a short birth interval ≤ 2 years  and newborns with a higher birth order with a longer birth interval > 2 years  were significantly associated with neonatal mortality. Other significant factors that affected neonatal deaths included neonates born to mothers younger than 20 years , neonates born to mothers residing in rural areas compared with urban residents , male neonates , mothers who perceived their neonate’s body size to be smaller than the average size , and mothers who delivered their neonates by caesarean section .Our study suggests that the Nigerian government needs to invest more in the healthcare system to ensure quality care for women and newborns. Community-based intervention is also required and should focus on child spacing, childbearing at a younger age, and poverty eradication programs, particularly in rural areas, to reduce avoidable neonatal deaths in Nigeria. Neonatal mortality is still a significant public health problem worldwide, and accounts for more than 60% of newborn deaths before their first birthday . Of the Globally, Nigeria ranks second to India with the highest number of neonatal deaths, with the highest reported number in Africa . Each yePrevious studies on neonatal mortality in Nigeria have indicated that low birth weight, lack of antenatal care, maternal illness, mother’s age, prematurity, and birth asphyxia are linked with neonatal mortality, but these studies were all hospital-based case–control and experimental studies -11. LimiThe main goal of this study was to determine factors associated with neonatal mortality using the 2008 NDHS. Findings from the study would be useful to public health researchers and policy makers in reviewing and designing new community based intervention strategies aimed at reducing neonatal mortality in Nigeria. Therefore, this study presents population-based data on risk factors associated with neonatal mortality in Nigeria.This study was based on a public domain dataset that is freely available online. The data were collected for the NDHS 2008 . The surThe 2008 NDHS was a stratified two-stage cluster design. Each state was stratified into two distinct groups of urban and rural areas. The census enumeration areas of the 2006 population census were used as the clusters for the 2008 NDHS. In the first stage, clusters were selected based on probability proportionate to the population size among its urban and rural areas. In each of the selected clusters, a complete listing of households was obtained. The listed households then served as the sampling frame for the selection of households to be interviewed in the second stage. Thereafter, a systematic sampling with equal probability was used in the second stage in selecting the specified number of households in each cluster for interview .A structured questionnaire was used for interviewing the selected households for the 2008 NDHS. The questionnaires that were administered to the respondent household members were the household questionnaire, the women’s questionnaire, and the men’s questionnaire. These questionnaires consisted of a series of questions on population and health issues. The household questionnaire recorded all of the usual residents of the selected household and their characteristics, such as age, sex, education, and their relationship with the head of the household, as well as information on amenities and features of the household’s dwelling unit. Additionally, the survey collected data on height and weight measurements for children aged younger than 5 years, and women aged 15–49 years. The women’s questionnaire consists of information included, but not limited to, birth history, childhood mortality, fertility preferences, knowledge and use of family planning methods, antenatal care, delivery, postnatal care, vaccinations, and childhood illnesses, as well as malaria prevention and treatment. The 2008 NDHS men’s questionnaire was the same as the women’s questionnaire, but did not contain a detailed reproductive history, maternal and child health, or nutrition. However, notably, gestational age, intrapartum-related complications, and birth asphyxia, which could potentially improve neonatal data, were not collected in the 2008 NDHS.A total of 888 clusters were selected for the 2008 NDHS sample survey. Of these clusters, a total of 36,298 households were selected for interview in the 2008 NDHS. At the time of the survey, nearly 5% of the households were not occupied. However, more than 98% of the occupied households were successfully interviewed. A total of 34,596 eligible women aged between 15 and 49 years were interviewed, yielding a response rate of 96.5%. The analysis was restricted to all singleton live births for a 5-year period preceding the 2008 NDHS to reduce recall bias about birth and death dates reported by mothers.A conceptual framework of child survival in developed and developing countries has been developed by other authors -17. HoweThe outcome variable for this study was neonatal death as reported by the mothers who participated in the survey, and it was defined as the death of a neonate between birth and 1 month of life. This takes a binary form, such that neonatal death will be regarded as a success (1 = if death occurs in the specified age period) or failure . The outcome variable was examined against all confounding variables, and these variables were classified into three distinct groups: community level factors, household factors, and individual level factors consisting of socioeconomic factors  was used instead of birth weight because a previous study showed that there is a close relationship between mean birth weight and perceived newborn size by the mother .The NMR was calculated by using a similar method described by Rutstien and Rojas . The cruThe multivariable analysis models conducted used a stepwise backwards elimination procedure to identify independent variables that were significantly associated with the study outcome. To avoid any statistical bias, we double checked our backward elimination method by using the following procedures: (1) we entered only potential risk factors with a p value < 0.20 obtained in the univariable analysis for backward elimination process, (2) we tested the backward elimination by including all of the variables , and (3) we tested and reported any collinearity in the final model. HRs and 95% confidence intervals (CIs) were calculated to assess the adjusted risk factors that affect study outcome, and those with p < 0.05 were retained in the final model.Table Neonates delivered by caesarean section had a higher NMR than those born vaginally (NMR: 89.9 vs 35.8). The NMR for male neonates was higher than that for female neonates (NMR: 41.4 vs 31.7).Newborns born to mothers residing in rural areas had a higher risk of neonatal mortality than those who lived in urban areas . Compared with neonates born to mothers aged between 30 and 39 years, neonates born to younger mothers (<20 years)  reported a significantly higher risk of neonatal deaths. When the place of residence was replaced by household wealth index in the final model, neonates born to mothers in poor households had a high risk of neonatal death, although this was not statistically significant .Male neonates were more likely  to die than female neonates in the first month of life. Neonates delivered by caesarean section had a significantly higher risk of neonatal mortality  compared with non-caesarean delivery. Neonates whose birth size were perceived by their mothers as small or smaller were also more likely to die than those of average or larger-sized neonates .As shown in Table The overall aim of this study was to identify risk factors associated with neonatal mortality in Nigeria using a nationally representative sample. This study showed several factors that were significantly associated with neonatal mortality after adjusting for confounding factors, and each of these factors are discussed below.We found that the NMR for singleton live-born infants between 2003 and 2008 was 36.7 (95% CI: 34.4–39.0). However, a preliminary report from the 2013 NDHS indicated that the NMR slightly fell by approximately 8% from 40 deaths per 1000 live births in 2008 to 37 in 2013 . DespiteOur study showed that male neonates had a significantly higher risk of dying during the neonatal period compared with female neonates. This finding is consistent with a cross-sectional study conducted in Indonesia in 2008, which indicated that male neonates were more likely to die than female neonates . AdditioIn our study, mothers who perceived the size of their newborns to be small or very small had a 2.26 times greater risk of dying in the first month of life than those mothers who perceived their neonates to be of average or larger size. Similarly, findings from a cross-sectional study conducted in five Asian countries  in 2008 also showed that smaller than average neonates had an increased risk of neonatal deaths than average or larger sized neonates in four of the five countries with data on perceived newborn size . Even thOur study showed that neonates delivered by caesarean section had a higher relative risk of neonatal mortality compared with vaginal deliveries. This result contradicts previous reports, which indicated a statistically insignificant relationship between the mode of delivery and neonatal mortality . A similThe current study observed that neonates born to mothers aged younger than 20 years had a significantly higher risk of mortality than those born to mothers aged 20–29 years, 30–39 years, and 40–49 years. This finding is similar to that reported in previous studies ,36. HoweWe found that children of birth order (2 through 4 or higher) born with a shorter birth interval (≤2 years) were at higher risk of dying than those with a longer birth interval (>2 years). This finding is similar to a cross-sectional study carried out in India, which showed that neonates of fourth or higher birth order with a shorter birth interval of < 2 years have an increased risk of death compared with those of second and third birth order with a longer birth interval of > 2 years . MaternaThe current study showed that neonates born to mothers residing in rural areas had a higher risk of neonatal mortality compared with those living in urban areas. This finding is consistent with previous studies ,41, whicThe strengths and weaknesses of this study need to be considered when drawing specific inferences. This study was a nationally representative survey, with a stratified two-stage cluster sampling design, which achieved a 98.3% response rate. Additionally, recall errors arising from dates of birth and death given by women interviewed in the survey were minimized by restricting our analyses to births within the 5-year period preceding the survey. Third, the proportion of missing data was relatively small, such that it may not have influenced findings in our study. Despite these strengths, a number of weaknesses were also present in the study and they are as follows. (1) Only surviving women were interviewed, which may have led to under-reporting of the number of newborn deaths because of the association of neonatal death with maternal death . (2) GesA community-based interventional study on reducing neonatal death in Nigeria should be performed to focus on using verbal autopsy and birth weight. To reduce the recall period of using these instruments, a verbal autopsy should be undertaken before the culturally prescribed mourning period . In addiOur analysis of factors associated with neonatal mortality in Nigeria revealed that living in rural areas, child bearing at a younger age, birth order and birth interval, sex of the newborn , caesarean delivery, and mothers who perceive their newborns as smaller than average at birth significantly increased the risk of neonatal death. Our findings indicate the need to implement community based newborn care interventions particularly, educating community health workers and traditional birth attendants about safe delivery practice, the benefits of Kangaroo mother care method on low birth weight newborns, child spacing and promote delay of first pregnancy will contribute to the improvement of neonatal mortality statistics in Nigeria.The authors declare that they have no competing interests.OKE and KEA were involved in the conception and design of this study. OKE carried out the analysis and drafted the manuscript. KEA, MJD, JH, and ANP gave advice on interpretation and revised and edited the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2458/14/521/prepub


Tumor treating fields (TTFields) are alternating electric fields frequency tuned to 200 kHz for the treatment of recurrent glioblastoma. We report a patient treated with TTFields and determined the distribution of TTFields intracranially by computerized simulation using co-registered postgadolinium T1-weighted, T2, and MP RAGE images together with pre-specified conductivity and relative permittivity values for various cerebral structures. The distribution of the electric fields within the brain is inhomogeneous. Higher field intensities were aggregated near the ventricles, particularly at the frontal and occipital horns. The recurred tumor was found distant from the primary glioblastoma and it was located at a site of relatively lower electric field intensity. Future improvement in TTFields treatment may need to take into account the inhomogeneity of the electric field distribution within the brain. We found that the electric fields on the scalp mirror that of the observed skin irritation. The strength of TTFields is highest at the ventricular horns and the medial surface of the tumor.Alternating electric fields or tumor treating fields (TTFields) therapy is a novel treatment modality for recurrent glioblastoma We performed an IRB-approved retrospective neuroimaging analysis of TTFields therapy in a 67-year-old woman with recurrent glioblastoma in the posterior right frontal brain Fig. and B 6 From the same baseline MRI used for array layout, a three-dimensional rendition of the head was generated based on the co-registered MR image datasets using ScanIP software . A finite element mesh was generated for each of the segmented head structures including the scalp, skull, dura, cerebrospinal fluid (CSF), supratentorial gray/white matter, ventricles, brainstem, cerebellum, and the recurrent glioblastoma. The composite finite element mesh model was imported into COMSOL Multiphysics 5.0  to solve for the electric field distribution. The respective conductivity and relative permittivity values used were 0.001 S/m and 1110 for scalp, 0.021 S/m and 204 for skull, 0.502 S/m and 290 for dura, 2.000 S/m and 109 for CSF, 0.141 S/m and 2010 for gray matter, 0.087 S/m and 1290 for white matter, and 1.000 S/m and 10,000 for recurrent glioblastoma This is the first electric field simulation of a patient with recurrent glioblastoma treated with TTFields therapy and our analysis revealed important information relevant to the clinical care of patients. First, the highest electric field strength was found around the ceramic disks forming each array Fig., which iE. T. Wong received an unrestricted grant from Novocure for sponsored laboratory research. E. Lok and V. Hua report no disclosure.


Nearly all women shower or take baths during early pregnancy; however, bathing habits  may be related to the risk of maternal hyperthermia and exposure to water disinfection byproducts, both of which are suspected to increase risk for multiple types of birth defects. Thus, we assessed the relationships between bathing habits during pregnancy and the risk for several nonsyndromic birth defects in offspring.Data for cases with one of 13 types of birth defects and controls from the National Birth Defects Prevention Study delivered during 2000–2007 were evaluated. Logistic regression analyses were conducted separately for each type of birth defect.There were few associations between shower frequency or bath frequency or length and risk for birth defects in offspring. The risk for gastroschisis in offspring was increased among women who reported showers lasting ≥15 compared to <15 minutes . In addition, we observed modest increases in the risk for spina bifida, cleft lip with or without cleft palate, and limb reduction defects in offspring of women who showered ≥15 compared to <15 minutes. The results of comparisons among more specific categories of shower length  were similar.Our findings suggest that shower length may be associated with gastroschisis, but the modest associations with other birth defects were not supported by analyses of bath length or bath or shower frequency. Given that showering for ≥15 minutes during pregnancy is very common, further evaluation of the relationship between maternal showering habits and birth defects in offspring is worthwhile. Approximately 76% of women in the U.S. shower at least once a day and approximately 28% of women in the U.S. take a bath at least once a day. AlthougIn mice, submersion in a hot bath has been shown to induce a variety of malformations (reviewed in). In epiIn animal models, exposure to water disinfection byproducts (WDBPs) has induced a variety of malformations (reviewed in). ShowerThe purpose of this study was to assess the relationship between shower and bath length and frequency during pregnancy and the risk for several nonsyndromic birth defects in offspring, using data available from the National Birth Defects Prevention Study.The National Birth Defects Prevention Study (NBDPS) is a large population-based case–control study of birth defects that includes ten surveillance sites . The Institutional Review Boards for each site approved the study protocol and the current analyses were approved by the Institutional Review Board at the University of Texas Health Science Center.The details of the NBDPS have been previously described. BrieflyData on exposures before and during pregnancy, family history of birth defects, maternal conditions, and lifestyle/behavioral factors were collected during a computer-assisted interview with participating mothers of cases and controls. Questions about bath and shower habits during pregnancy were added to the interview around the year 2000. Specifically, mothers were asked how often they showered during pregnancy, as well as how many minutes were typically spent per shower. These questions were also repeated for baths. Women were also asked if they left the window open or turned on the exhaust fan when showering or taking baths.The present analyses were conducted among subjects with estimated dates of delivery from January 1, 2000 through December 31, 2007. Cases with any of the following non-cardiac birth defect phenotypes, which have previously been associated with maternal exposure to hyperthermia or WDBPs, were included: anencephaly, spina bifida, cleft lip with or without cleft palate, cleft palate without cleft lip, microphthalmia , congenital cataract, esophageal atresia, gastroschisis, omphalocele, bilateral renal agenesis or hypoplasia, diaphragmatic hernia, limb reduction defects, and hypospadias.Shower and bath length were categorized as <15 minutes or ≥15 minutes per occurrence, based on a median split of shower length among controls. The distributions of the following variables were compared across these two categories of shower length among controls: maternal race/ethnicity, age at delivery, education, surveillance site, body mass index, daily folic acid use (during the month before conception through the first month of pregnancy), nulliparity, smoking (during the month before pregnancy through the first trimester), season of conception, and annual household income. Data for these variables were collected during the computer-assisted telephone interview. Distributions of these variables across categories of bath length, bath frequency, and shower frequency were not compared because there was less variability in these bathing habits .a priori potential confounders (selected based on previous literature): study site and maternal age at delivery, body mass index, education, race/ethnicity, income, parity, folic acid use, smoking, and season of conception. All analyses of hypospadias were restricted to male control infants.All analyses were conducted separately for each phenotype. Infants with more than one birth defect were included in analyses of each birth defect. Unconditional logistic regression was used to assess the crude relationship between shower and bath length and frequency and risk for each birth defect. Main adjusted analyses were conducted for each birth defect, adjusting for the following Several additional analyses were conducted for shower length but not bath length, as the majority of women reported not taking baths (see Results). Main adjusted analyses of shower length were repeated again, stratified by presence of steam exhaust (use of an exhaust fan or leaving a window open during showers), because these activities are expected to decrease exposure to both hyperthermia and WDBPs. The adjusted analyses for shower length were also repeated in separate models, further adjusting for shower frequency , bath frequency , and bath length . To further assess the possibility that risk may vary among the offspring of women who took longer showers, the main adjusted analyses were repeated based on more specific categories of longer shower length .post-hoc exploratory analyses where performed to evaluate the relationship between shower length and gastroschisis within specific maternal subgroups . The main shower length analyses were also repeated among subjects that did not take baths .Because longer shower length was moderately associated with gastroschisis (see Results), and because previous literature suggests that the effects of gastroschisis risk factors may vary within specific subgroups ,18, addiOf 6,758 total control mothers, we observed that 95.6%  reported taking showers at least once per month during pregnancy and 50.4%  reported taking showers that lasted at least 15 minutes. Characteristics of control mothers were compared between women that took <15 minute and ≥15 minute showers  . Specifically, our results suggest that longer showers may modestly increase risk for anencephaly, spina bifida, cleft lip with or without cleft palate, gastroschisis, renal agenesis, and limb reduction defects. The exact mechanism that may be responsible for these associations is unknown. Two plausible explanations include maternal exposure to water disinfection by-products and maternal hyperthermia, but it is not possible to differentiate between these mechanisms in our data and more research is needed to better understand how shower and bath length and frequency affect these exposure in humans.One of the strongest and most consistent associations we observed was between shower length and gastroschisis. It has been shown that hyperthermic exposure can induce gastroschisis in chick embryos. For example, gastroschisis was present in 100% of 33 embryos that survived incubation at 41 degrees C (approximately 106 degrees F) and in nearly 92% of 54 that did not survive. Few stuInterest in understanding maternal hyperthermic exposures and birth defect risk is growing, as this mechanism is increasingly suspected to be a risk factor for a broad range of birth defects. In addition to water-related hyperthermic exposures, additional sources of interest include maternal fever, sauna use, ambient temperature, electric blankets, ultrasound and electromagnetic radiation, occupation-related hyperthermia, and medication-induced temperature increases. Although maternal hyperthermia has been evaluated relatively extensively for neural tube defects  and is strongly suspected to increase risk for neural tube defects (reviewed in), hypertToxicants in tap water, such as WDBPs, are also emerging as suspected risk factors for a wide range of birth defects. In addition to gastroschisis, WDBPs have been associated with neural tube defects, but do not appear to be associated with cleft lip (reviewed in). Other Results from our evaluations of bath length and shower and bath frequency were less consistent than our main findings and did not support our shower length findings for most birth defects; however, because most women took one shower per day (68% of controls) and no baths (54% of controls), it is likely that these analyses were limited by the small amount of variability.This study serves as a first step to understanding the relationship between shower length and birth defect risk and should be considered in light of potential limitations. Although we adjusted for study center, we were unable to directly account for variability in ambient temperature, water temperature or contaminants, maternal body temperature, or additional potential household water exposures . Furthermore, it is possible that there may have been some misclassification in our main exposure variables, as shower length was self-reported, potentially several months after pregnancy. However, showering length is somewhat habitual and may thus be relatively stable over the course of several years. Additionally, it may be difficult to measure exposure more accurately than self-report  in studies of rare outcomes, such as birth defects , so these analyses based on self-reported shower length serve as an important first step. Direct measurements of maternal hyperthermia or exposure to water disinfection byproducts were unavailable, but this study has provided some rationale for follow-up studies to assess showering habits in more detail.There were many strengths of this study, including use of data from one of the largest studies of birth defects in the world. Our use of a population-based sample that included fetal deaths and elective pregnancy terminations likely limited the influence of potential selection bias. Further, using a sample of nonsyndromic cases likely reduced case heterogeneity. We also conducted several sensitivity analyses and supplemental analyses that support our main findings.The findings reported here serve as preliminary evidence suggesting that ≥15 minute showers during pregnancy could have potential teratogenic effects, particularly for gastroschisis. Given the high prevalence of ≥15 minute showers , additional studies are warranted to replicate our findings, better understand the mechanisms that may be involved, and evaluate the public health implications.CI: Confidence interval; OR: Odds ratio; NBDPS: National birth defects prevention study; WDBPs: Water disinfection byproducts.The authors declare that they have no competing interests.AJA: Conception and design of the study, analysis of the data and writing of the manuscript. LEM, PJL, MAC, DKW: Contributed to conception and design of the study, interpretation of analysis and writing of the manuscript. All five authors read and approved this version of the manuscript.Table S2. Crude and adjusted odds ratios for the associations between average bath length and risk for birth defects, National Birth Defects Prevention Study, USA, 2000-2007. Table S3. Crude and adjusted odds ratios for the associations between number of showers or baths per day and risk for birth defects, National Birth Defects Prevention Study, USA, 2000-2007. Table S4. Associations between average shower length and risk for birth defects adjusted for shower frequency and bath frequency and length, National Birth Defects Prevention Study, 2000-2007. Table S5. Crude and adjusted odds ratios for the associations between average shower length and risk for birth defects, National Birth Defects Prevention Study, USA, 2000-2007. Table S6. Adjusted odds ratios for the associations between average shower length and risk for birth defects by presence of steam exhaust, National Birth Defects Prevention Study, USA, 2000-2007. Table S7. Crude and adjusted odds ratios for the associations between average shower length and risk for birth defects among participants who did not take baths, National Birth Defects Prevention Study, USA, 2000-2007.Shower and bath characteristics among control mothers, National Birth Defects Prevention Study, USA, 2000-2007. Click here for file


When we conducted a stratified analysis according to study design, the risk estimate was slightly weaker in cohort studies  than in cross-sectional studies . Additionally, the associations between passive maternal smoking and preterm birth were statistically significant for studies conducted in Asia , for studies including more than 100 cases of preterm birth , and for studies adjusted for maternal age , socioeconomic status and/or education , body mass index , and parity . Our findings demonstrate that passive maternal smoking is associated with an increased risk of preterm birth. Future prospective cohort studies are warranted to provide more detailed results stratified by passive maternal smoking during different trimesters of pregnancy and by different types and causes of preterm birth.Previous studies investigating the relationship between passive maternal smoking and preterm birth reveal inconsistent results. We conducted the current meta-analysis of observational studies to evaluate the relationship between passive maternal smoking and preterm birth. We identified relevant studies by searching PubMed, EMBASE, and ISI Web of Science databases. We used random-effects models to estimate summary odds ratios (SORs) and 95% confidence intervals (CIs) for aforementioned association. For the analysis, we included 24 studies that involved a total of 5607 women who experienced preterm birth. Overall, the SORs of preterm birth for women who were ever exposed to passive smoking  Preterm birth, which is birth before 37 completed weeks of gestation, is a leading cause of neonatal death worldwide. Approximately 15% of preterm infants die within one month after birth –3. ThereActive smoking is well-established as a contributing factor to preterm birth. There is increasing scientific and regulatory concern for the role that passive smoking, which is the exposure to environmental tobacco smoke or second-hand smoke, may play in preterm birth, possibly due to the same biological mechanisms as active smoking –5. HowevA large portion of women in the general population are exposed to passive smoking , so evenWe performed a comprehensive search of articles published through February 28, 2015 by searching PubMed, EMBASE, and Web of Science databases. The following terms were used in the electronic search:  and . We also manually searched the references cited in the retrieved articles. This meta-analysis was planned, conducted, and reported in adherence with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines .Studies were selected and excluded by 2 independent investigators (Q-JW and T-TG). Published articles were included according to these selection criteria: 1) the study used an observational study design ; 2) the study provided information on passive maternal smoking  as the exposure; 3) the study reported preterm birth  as the outcome; and 4) the study reported usable risk estimates between passive maternal smoking and preterm birth.Published articles were excluded according to the following criteria: 1) the study was a review without original data, an ecological study, an editorial, or a case report; 2) the study reported the risk estimates for the highest category relative to the lowest category of passive maternal smoking instead of exposure to passive maternal smoking; and 3) the study investigated passive maternal smoking in a certain trimester of pregnancy instead of the entire pregnancy.Two independent reviewers (Q-JW and T-TG) completed the data extraction using a predefined sheet. Dissimilarities were resolved by discussion between the authors. The following data were extracted from each included study: first author’s name, year of publication, country of study, study design, period of exposure measurement, number of subjects with preterm birth and sample size of the study, categories of exposure with corresponding risk estimates, and potential confounders adjusted in the primary analysis.We used the Newcastle-Ottawa Scale (NOS) –21 and AQ and I2 statistics. For the Q statistic, a P-value less than 0.1 was considered to represent statistically significant heterogeneity. For the I2 statistic, a value greater than 75% was considered to indicate significant heterogeneity; a value less than 25% indicated the absence of significant heterogeneity [Since the majority of included studies reported risk estimates as odds ratios (ORs) –14,26–36ogeneity ,47. We sTo find the possible sources of heterogeneity of the primary results, we carried out the stratified analyses according to the following study features for all studies: design of study , study location , median number of cases , time of exposure measurement (before delivery and after delivery), and potential confounders adjusted in the analyses . Small study bias was assessed by visual inspection of a funnel plot ,48 and bThe detailed article screening processes are outlined in The characteristics of the 24 articles are described in Characteristics related to study quality are summarized in Tables I2 = 36.1%, P = 0.038) (P = 0.51) or Begg's test (P = 0.91), and no asymmetry was noted in the funnel plot upon visual inspection . We found no significant association between passive smoking at work or another place and preterm birth.Overall, compared to women who were never exposed to passive smoking during pregnancy, women who had ever been exposed had a significantly increased risk of preterm birth , with moderate heterogeneity (= 0.038) . Publicaspection . When thspection , a signiI2 = 0%), 1.47 , and 1.45 , respectively  after omission of the study by Khader et al [I2 = 33.9%) after omission of the study by Kim et al [I2 = 42.6%). Lastly, we excluded 8 studies [I2 = 50.9%).er et al  to 1.24 im et al . Additioim et al  in whichim et al ; this re studies ,36,41–44The rate of preterm birth has increased in most countries in the past decade and it represents an important public health issue. A previous meta-analysis of studies published prior to 2009 found no significant association between passive maternal smoking and risk of preterm birth . For thiWhen passive smoking and preterm birth were examined according to exposure location, a statistically significant association was found only for passive smoking exposure at home . Since oIn the subgroup analyses stratified by study design, we found that the point estimate of the relationship between preterm birth and passive smoking was slightly stronger among cross-sectional studies and weaker among cohort studies. Compared with cross-sectional or retrospective studies, prospective studies had fewer biases due to their prospective nature. However, we observed non-significant associations between passive smoking and preterm birth among studies that collected exposure information before delivery. This same pattern was observed when we restricted the stratified analysis to cohort studies . This isA significant association between passive maternal smoking at any place and preterm birth risk was only observed in studies from Asia , which cSeveral potential biological mechanisms have been suggested to explain the positive association between passive maternal smoking and preterm birth. Passive smoke contains several toxic chemicals, including nicotine, carbon monoxide, and DNA adducts –55. PrevOur current meta-analysis has several strengths. First, to the best of our knowledge, this is the most comprehensive and current meta-analysis for evaluating the association between passive maternal smoking and preterm birth. Second, our meta-analysis included 24 observational studies, 16 of which were cohort studies, that involved 5607 patients from a total population of approximately 88,200 participants; this provided sufficient power to detect modest associations. Third, compared with previous meta-analyses , we condSeveral potential limitations of this meta-analysis should be acknowledged. First, considering the nature of observational studies, we could not fully rule out the possibility of residual confounding. When we assessed the quality of the included studies, only 5 cohort studies ,27,32,52In conclusion, in this updated and comprehensive meta-analysis, we found that women who had ever been exposed to passive maternal smoking at any place or at home had a significantly higher risk of preterm birth than women who had never been exposed to passive smoking. Future prospective cohort studies are warranted to examine potential confounders of this association and to provide more detailed results that are stratified by passive maternal smoking in different trimesters and by different types and causes of preterm birth.S1 Checklist(DOC)Click here for additional data file.S1 Database(XLSX)Click here for additional data file.


Studies on birth weight and congenital anomalies in sub-Saharan regions are scarce.Data on child variables  and maternal variables  were collected for all neonates delivered at Ayder referral and Mekelle hospitals (Northern Ehthiopia) in a prospective study between 01-12-2011 and 01-05-2012.The total number of deliveries was 1516. More female (54%) than male neonates were born. Birth weights were 700-1,000 grams between 26 and 36 weeks of pregnancy and then increased linearly to 3,500-4,000 grams at 40 weeks. Thirty-five and 54% of neonates were very-low and low birth weight, respectively, without sex difference. Very-low birth-weight prevalence was not affected by parity. Male and female neonates from parity-2 and parity-2-4 mothers, respectively, were least frequently under weight. Sixty percent of newborns to parity -3 mothers weighed less than 2,500 grams, without sex difference. The percentage male neonates dropped from ~50% in parity-1-3 mothers to ~20% in parity-6 mothers. Diagnosed congenital malformations (~2%) were 2-fold more frequent in boys than girls. The commonest malformations were in the central nervous system . Parity, low birth weight, gestational age less than 35 weeks, male sex, and lack of antenatal care were the most significant risk factors for congenital anomalies.The high prevalence of neonates with low birth weight and CNS anomalies in Northern Ethiopia was very high. The findings may reflect the harsh conditions in the past 2 decades and suggest environmental and/or nutritional causes. Male sex and parity affected the outcome of pregnancy negatively. Birth weight is a technically simple parameter to monitor prenatal health in a population. Establishing the prevalence of low birth weight (LBW) is particularly important, since perinatal morbidity and mortality are more frequent in LBW than in normal infants and has become the second cause of death in this period, after premature birth . It has The number and especially the type of congenital malformations in newborns can also reflect prenatal health in a population. The global prevalence of all congenital malformations is 2-3% , while tAlthough underreporting, deficiencies in diagnostic capabilities, and poor follow-up at birth may affect the report of congenital malformations , the actNo prospective study on birth weight and the incidence of congenital anomalies is, to our knowledge, available for Northern Ethiopia. We, therefore, started such a study to obtain a first impression of neonatal health at birth, the frequency of congenital anomalies, and potential risk factors in the Ayder refferral and Mekelle zonal hospitals. Ayder referral hospital is a teaching hospital under Mekelle University and administered by the Ministry of Education. The hospital has 457 beds and provides full medical services. The hospital labor ward has 6 waiting beds, 3 couches and currently sees 1,000-1,100 deliveries per year. Mekelle zonal hospital is the oldest hospital in Mekelle city and administered by the Ministry of Health. It has 240 beds and a labor ward with 2 waiting beds, 3 couches and currently around 2,400 deliveries per year. Though the estimated number of deliveries in the town and surrounding district Enderta in the study period was ~6,000, only 1,516 (~25%) deliveries were attended in both hospitals. In general, ~6% deliveries in Ethiopia are attended in health institutions [This prospective cohort study was conducted between 01-12-2011 and 31-05-2012 and included all deliveries. Data on child variables  and maternal variables  were collected at Ayder referral and Mekelle hospitals. Birth weight, gestational age, gravidity, parity and maternal age were considered as continuous variables, while sex, antenatal care, previous abortions, maternal illness, medication, malformation history were categorical. Births outside the two hospitals were excluded. The births were attended by midwives and midlevel obstetric professionals (trained in emergency surgery and obstetrics). Birth weight was measured using baby weighing scale pan type in which the weight was estimated to the nearest 100grams. Neonates weighing <1,500 and <2,500 grams were considered very low birth weight (VLBW) and low birth weight (LBW), respectively. Assesment of gestational age was deduced from the last reported menstrual period. Ultrasound examinations were not carried out routinely, because the equipment was not available in the antenatal care centers. All relevant information about birth weight and the presence of gross congenital defects detected within 24 hours after delivery was collected. The data was entered into a predesigned data collection format. The distribution of gestational age, parity, birth weight and sex was visualized as frequencies or proportions. The association of malformations with child and maternal variables was analyzed with classical logistic regression including the confounders, using Stata version 13. A P value of <0.05 was considered significant and <0.10 as indicating a trend. The ethical clearance for the study was obtained from Mekelle University-College of Health Sciences, Health Research Ethics Review Committee (HRERC) within Research and Community Service Team. Verbal informed consent was obtained from the mothers, as most of them were illiterate, prior to the beginning of the interviews.The total number of deliveries in the two hospitals during the 6-months study period was 1,516. Of these, 1484 had no diagnosed malformation Table A). The pχ2 test)). The female newborns were all born before 36 weeks, whereas the 20 out of the 22 male newborns (~90%) were all born after 36 weeks gestation  had visible congenital anomalies. The gestational age at which these babies were born is shown in Figure n Figure D. Male gTable Table The present study of a 6-months cohort of newborns in Northern Ethiopia reveals that newborns hardly increased in birth weight if born between 26 and 36 weeks, but rapidly thereafter. Thirty-five % of the cohort had a birth weight of <1,500 grams and 54% a birth weight of <2,500 grams. The prevalence of diagnosed malformations was 2.1%, but difficult to judge as only externally visible malformations were included. The number of CNS malformations was, at 1.5% of all births, very high. Males appeared to be more affected by the prevailing conditions in Northern Ethiopia than females: the offspring sex ratio was lower than reported elsewhere  and declined dramatically to only ~20% males if the neonates were born to mothers with at least 3 completed pregnancies, which, as far as we are aware, was not reported earlier. Furthermore, malformations were more often seen in males than females (~60%). We estimated that ~25% of the pregnant women from the region of Mekelle and Enderta district had their delivery supervised in one of both hospitals, which leaves ample room for selection. One indication for selection is the higher prevalence of malformations in Ayder referral hospital than in Mekelle hospital. Antenatal care in Northern Ethiopia does, on the other hand, not yet include advanced diagnostic tools, such as an ultrasound examination. For this reason, we do not think that the population of women who delivered in the hospital was strongly biased by selection. The indices, therefore, show that the study reports an underpriviledged population, even by regional standards, and provides intriguing clues about the characteristics of the population, the risk factors that may lead to fetal congenital anomalies, and the pattern and occurrence of malformations in Northern Ethiopia. Thus far, this kind of study was nonexistent in Ethiopia, even though the death rate from congenital anomalies is believed to be high .The birth weight of the sample population hardly increased between 26 and 36 weeks gestation and increased only thereafter. This relation between birth weight and increasing gestational age at birth is exceptional and best resembled the lowest percentile of the reference curves of neonates in a Western country, such as The Netherlands . HoweverThe very high prevalence of severely underweight newborns after 3-4 prior deliveries (~60%) in our cohort is also noticeable. For Harare, Zimbabwe, a prevalence of ~37% of small-for-gestational-age (SGA) infants was reported . ElsewheThe incidence of malformations was ~3.8% in Ayder referral hospital and 1.3% in Mekelle hospital. This result probably underestimates the actual percentage, because some of the major anomalies, such as cardiac defects with a prevalence of 12 per 1,000 live births in the Western world , were noThe most frequently observed malformations concerned the central nervous system , with the majority being spina bifida. This is probably due to the very visible presentation of these defects and, therefore, probably a parameter for less obvious malformations as well. In a similar study in Cameroon, 0.2% of the neonates suffered from CNS defects, of which spina bifida was the most prevalent with 73% . In contThe population of newborns with VLBW and LBW in Northern Ethiopia was high in 2012. Likewise, the percentage of newborns with CNS malformations was high. Male newborns appear more sensitive to the adverse conditions in Northern Ethiopia than female newborns.


Preterm birth (PB) and fetal growth restriction (FGR) convey the highest risk of perinatal mortality and morbidity, as well as increasing the chance of developing chronic disease in later life. Identifying early in pregnancy the unfavourable maternal conditions that can predict poor birth outcomes could help their prevention and management. Here we used an exploratory metabolic profiling approach (metabolomics) to investigate the association between birth outcomes and metabolites in maternal urine collected early in pregnancy as part of the prospective mother–child cohort Rhea study. Metabolomic techniques can simultaneously capture information about genotype and its interaction with the accumulated exposures experienced by an individual from their diet, environment, physical activity or disease (the exposome). As metabolic syndrome has previously been shown to be associated with PB in this cohort, we sought to gain further insight into PB-linked metabolic phenotypes and to define new predictive biomarkers.Our study was a case–control study nested within the Rhea cohort. Major metabolites (n = 34) in maternal urine samples collected at the end of the first trimester (n = 438) were measured using proton nuclear magnetic resonance spectroscopy. In addition to PB, we used FGR in weight and small for gestational age as study endpoints.N-acetyl resonance (IOR = 5.84 CI 1.44 to 39.50). This resonance was positively correlated with body mass index, and stratified analysis confirmed that N-acetyl glycoprotein and IPB were significantly associated in overweight and obese women only. Spontaneous PB cases were associated with elevated urinary lysine (IOR = 2.79 CI 1.20 to 6.98) and lower formate levels (IOR = 0.42 CI 0.19 to 0.94).We observed significant associations between FGR and decreased urinary acetate (interquartile odds ratio (IOR) = 0.18 CI 0.04 to 0.60), formate (IOR = 0.24 CI 0.07 to 0.71), tyrosine (IOR = 0.27 CI 0.08 to 0.81) and trimethylamine (IOR = 0.14 CI 0.04 to 0.40) adjusting for maternal education, maternal age, parity, and smoking during pregnancy. These metabolites were inversely correlated with blood insulin. Women with clinically induced PB (IPB) had a significant increase in a glycoprotein Urinary metabolites measured at the end of the first trimester are associated with increased risk of negative birth outcomes, and provide novel information about the possible mechanisms leading to adverse pregnancies in the Rhea cohort. This study emphasizes the potential of metabolic profiling of urine as a means to identify novel non-invasive biomarkers of PB and FGR risk. Fetal growth restriction (FGR) and preterm birth (PB) are the main predictors of perinatal morbidity and mortality ,2. TheseA recent report evocated a sharp increase in late PB in Greece for the past 20 years, in a similar fashion to what has been noted in other middle or high income countries, potentially associated with increased maternal age and a change in obstetric interventions . Other fThe study was conducted according to the guidelines laid down in the Declaration of Helsinki, and all procedures involving human subjects were approved by the ethical committee of the University Hospital in Heraklion, Crete, Greece. Written informed consent was obtained from all women participating in the study.The Rhea project is a mother–child study, prospectively examining a population-based cohort of pregnant women and their children in the prefecture of Heraklion, Crete, Greece . Female 1H-NMR) spectra were acquired, of which 26 spectra were excluded (because of high dilution or high excretion of drug metabolites), leaving 438 spectra available for modelling the metabolite profile with respect to birth outcome.During this study period 1,317 women were followed up until delivery. Women with incomplete diagnostic information, multiple pregnancies, diagnosed pre-eclampsia (a condition associated with PB), spontaneous or induced abortion, or who gave birth to stillborn infants were not included in the study . Our metPB, the primary outcome of interest, is defined as premature delivery at less than 37 weeks of gestation . The gesA multivariable fractional polynomial linear regression model was used to predict birth weight, allowing polynomial terms for continuous variables in the linear regression models. The final model included as covariates the gestational age, infant gender, maternal and paternal height, pre-pregnancy maternal weight, and interaction of gestational age with maternal weight. Gestational age and type of PB were known for 438 women, whereas FGR data were available for only 401 women because a number values necessary to define the outcome were missing.Data on plasma triglycerides, total cholesterol, high density lipoprotein cholesterol (HDL-C) and low density lipoprotein cholesterol (LDL-C) of 227 fasting pregnant women at the first prenatal visit were available . The ins2HPO4/NaH2PO4, pH 7.4) to minimise variations in chemical shift values in the acquired 1H NMR spectra due to minor pH differences. This buffer contained 1 mM sodium 3-trimethylsilyl-(2H4)-1-propionate (TSP) in 20% D2O and 3 mM of the bacteriostatic agent sodium azide (NaN3). TSP is a chemical shift reference (δ = 0.00) and D2O provided a field-frequency lock. The buffered urine sample was then centrifuged at 16,000 × g for 5 minutes to remove any debris, and 550 μL of the resulting supernatant was pipetted into standard 5 mm NMR tubes [Urine samples were stored at −80 °C until analysis. An aliquot of 400 μL of urine was added to 200 μL phosphate buffer solution  operating at 600.13 MHz. The 1H NMR spectra of the urine samples were acquired using a standard one-dimensional pulse sequence with water pre-saturation  during both the recycle delay (2 seconds) and mixing time . The 90° pulse length was adjusted to approximately 10 μs and t1 was set to 3 microseconds. For each sample, 128 free induction decays (FIDs) were collected into 32 K data points using a spectral width of 12,000 Hz. The FIDs were multiplied by an exponential weighting function corresponding to a line broadening of 0.3 Hz prior to Fourier transformation [ormation .All NMR spectra  were imported into MATLAB 7.3.1 (MathWorks), and were referenced and corrected for phase and baseline distortion using an in-house script . The spectral region δ 4.0 to 5.4, containing residual water and urea resonances, were removed prior to median fold change normalisation . Integra1H NMR spectroscopic signals were assigned to metabolites after reference to the literature [1H-1H correlation spectroscopy and 1H-1H total correlation spectroscopy.terature ,31 or onterature , and/or 2 test. The threshold statistical significance was set at a P < 0.05 and conducted with a two-sided alternative hypothesis.All statistical analyses were performed using R project software . ContinuP value cut-off) [2 test) for trend in proportions, was used to assess the frequency distribution of women with pregnancy outcomes according to the quartiles of the metabolites [Statistical analyses were conducted on 34 metabolites to assess their variation in relation to birth outcomes  and to maternal parameters . A five-step analysis was conducted to select metabolites that were significantly associated with birth outcomes and associated with metabolic syndrome. To identify metabolites associated with birth outcomes, a non-parametric test (Mann–Whitney U–test), was used, because of the non-normal distribution of the metabolite relative concentrations. The effect of multiple testing was considered by calculating the false discovery rate , Spearman’s correlation coefficients were calculated. Metabolites with significant association with birth outcomes in logistic regression models and significant correlation coefficients with metabolic syndrome traits were selected for the final analysis. A stratified analysis by maternal BMI before pregnancy and maternal insulin levels at the first prenatal visit, was performed using multivariate logistic regression models on log-transformed metabolite levels, correcting for potential confounders (as described above).Our metabolomics study was designed as a case–control study nested within the Rhea cohort. Table 1H NMR spectroscopy was applied to all the urine specimens from our study population (n = 464). From these spectra, 26 were excluded because of high dilution or high excretion of drug metabolites, leaving 438 available for modelling the metabolite profile with respect to birth outcomes. In total, 34 metabolites were identified in the urinary 1H NMR spectra ; amino acids such as alanine, leucine and tyrosine; and other metabolites such as p-cresol sulphate and niacin metabolites (N-methyl-2-pyridone-5-carboxamide or 2-Py).To obtain metabolic profiles, N-acetyl neuraminic acid.A systematic analysis was performed to detect associations between birth outcomes  and metabolite abundance. Because a single molecular species may give rise to multiple resonances (peaks) in an NMR spectrum, we chose to select a single representative peak for each metabolite  to provide the measurement, with most metabolites exhibiting well-resolved peaks analysed by spectral integration. Our strategy was to use two different univariate approaches for initial candidate selection, and to look for agreement between these to define a consensus set of metabolites. These candidate metabolites were then subject to multivariate regression analysis to control for major confounding in our study. The first selection approach tested for significant median differences in metabolite abundance between cases and controls for each outcome. For a full description of the results of this analysis and the integral regions used for all metabolites, see Additional file P < 0.05) median differences between PB and control cases , glycine, TMAO, lysine and the singlet at 0.63 ppm significantly varied between SPB and control groups. The IPB group exhibited specifically higher levels of N-acetyl glycoproteins and lower levels of phenylacetylglutamine compared with controls. Using FDR analysis, we also estimated the likelihood of each difference between the groups being a false positive association  and separately on each subtype (SPB and IPB). Formate and an unassigned singlet resonance at 0.63 parts per million (ppm) probably derived from a steroid moiety, displayed significant . Out of the eight candidate metabolites for FGR and the two metabolites applying to both PB and IPB, all showed a trend in frequency of birth outcomes across quartiles, therefore showing a dose–response relationship between levels of candidate metabolites and the outcome incidence. However, only three of the six candidate metabolites for SPB, namely formate, lysine and the singlet at 0.63 ppm, showed a significant trend . High lysine and low formate levels were significantly associated with a higher risk of SPB.Finally, risk estimates of pregnancy outcomes were then computed using candidate metabolites as predictors in a logistic regression model, allowing adjustment for confounding factors such as maternal education, maternal age, parity and smoking habits Table . The IORN-acetyl glycoproteins in IPB cases, were associated with a steep increase in outcome incidence only at high level. ROC analysis was also performed on the metabolites that were significantly associated in adjusted logistic regression models, in order to provide an alternative test of the ability of these molecules to predict birth outcomes in the study population. Modest but statistically significant area under the curve (AUC) values were obtained for all metabolites . Increased BMI was associated with elevated levels of N-acetyl glycoprotein fragments in the urine (ρ = 0.14). BP was poorly correlated with urinary metabolites. These findings suggest that some of the variation in the urinary metabolites associated withbirth outcomes could be related to underlying maternal metabolic disease such as obesity and insulin resistance. Stratified analysis by maternal BMI as the two categories of ‘underweight and normal’ (<25) versus ‘overweight and obese’ (>25), confirmed that N-acetyl glycoprotein and IPB are significantly associated in overweight and obese women only . Figure N-acetyl glycoprotein levels were particularly high in IPB women with high BMI before pregnancy. A stratified analysis was also performed for insulin levels (low levels ≤6 mU/mL versus high levels >6 mU/mL). Tyrosine, acetate and formate associations with FGR were not significant in the high insulin group.According to a previous analysis , the pree ρ = 0.1. BP was 1H NMR-based metabolic profiling approach, we found early (end of first trimester) differences in urinary metabolic phenotypes in the pregnant women in the Rhea cohort study in whom PB and FGR subsequently occurred. These potentially predictive metabolic signatures of birth outcomes were correlated with aspects of metabolic syndrome. Furthermore, we observed a distinction between the metabolic signature of ‘medically indicated’/induced and ‘non-indicated’/spontaneous PB, suggesting a range of aetiological metabolic factors contributing to PB.Although over 90% of fetal growth occurs in the second half of gestation, maternal metabolism in the first trimester undergoes extensive changes in lipid storage, nitrogen species excretion and other metabolic pathways in order to facilitate fetal development . Thus, eN-acetyl glycoprotein fragments was observed in these women. The N-acetyl proton resonances arising at δ 2.04 ppm are frequently associated with inflammation-induced acute phase proteins such as alpha-1 glycoprotein when reported in serum [N-acetyl glycoprotein resonance was positively correlated in this study with BMI. In the Rhea cohort, pregnant women with metabolic syndrome (and in particular the obesity component) had a high risk of encountering IPB . The mechanisms relating N-acetyl glycoproteins to obesity and IPB, remain unclear; however, it is widely reported that high levels of adipose tissue can lead to systemic inflammation through release of cytokines such as interleukin-6, which could lead to an acute phase response [N-acetyl glycoprotein levels in NMR spectra of women with PB were also found in a study profiling maternal serum and cord blood at birth [Despite the relative small number of induced preterm pregnancies (n = 28), a significant increase in in serum , but thein serum . The N-aresponse . Higher at birth . Spontanat birth . The steat birth ,44. In o. The FGR metabolite profile was broadly inversely associated with plasma insulin and positively correlated with HDL-C levels. Of these metabolites, formate, tyrosine and trimethylamine were all found to be significantly positively correlated with each other, suggesting a common source of variation (ρ(formate-tyrosine) = 0.38, ρ(formate-trimethylamine) = 0.21 and ρ(tyrosine-trimethylamine) = 0.26). Elevations of several of these metabolites in blood have been previously associated with risk of insulin resistance [P = 0.23).With the exception of formate, a different metabolic profile  was associated specifically with poorer fetal growth. FGR remains difficult to assign owing to healthy biological variability in human size, hence in this study we used a definition based on customised birth-weight percentiles designed to better differentiate between infants who are small because of restricted growth and infants who are small but have reached their likely individual growth potential (see Methods) ,26. A sisistance ,46; howesistance  and intesistance . Howeversistance -51, hencsistance . This insistance . HoweverDespite our study not being directly comparable with previous metabolomics studies investigating birth outcomes, owing to differences in the analytical platform used and the biofluids studied (often cord blood serum or amniotic fluid), and because our samples were taken at the end of the first trimester (whereas most previous metabolomics studies have examined late pregnancy samples), some similarities with other investigations were observed. In addition to the instances cited above, a previous study in women with subsequent FGR also reported decreased levels of urinary trimethylamine, tyrosine and glycine . HoweverOur study has a number of other important limitations. Firstly, our study was not specifically designed to examine FGR, and only a limited number of these cases were available within our dataset. Secondly, although our study is unique in defining associations between metabolism during early pregnancy and birth outcomes, it is not possible at this stage to distinguish between pregnancy-induced effects and underlying metabolic risk factors. However, this does not negate the potential value of urinary metabolites in general as biomarkers of risk of negative birth outcomes, and our exploratory study has generated several hypotheses for future investigation. It is also possible that our observations reflect aetiological factors specific to the Rhea cohort, which experience an abnormally high rate of PB, and are not generalisable to the broader European population. Specimens from an independent cohort would be needed to validate our findings, and several efforts to complete comparable studies in large birth cohorts are currently underway, such as the Human Early-Life Exposome (HELIX) project . This coUrinary acetate, tyrosine, formate, trimethylamine, lysine and glycoprotein measured at the end of the first trimester are associated with increased risk of negative birth outcomes in the Rhea cohort. We believe our study to be a confirmation of the potential of metabolomics to reveal novel links between metabolite exposure and birth outcomes, and evidence in support of the inclusion of such approaches in studies that attempt to link the exposome to neonatal health.N-methyl-2-pyridone-5-carboxamide; FGR: Fetal weight growth restricted; IPB: Induced preterm birth; NMR: Nuclear magnetic resonance; PB: Preterm birth; SPB: Spontaneous preterm birth. BMI body mass index; BP: blood pressure; HDL-C: high density lipoprotein cholesterol; LDL-C: low density lipoprotein cholesterol; SGA: small for gestational age; PB: spontaneous preterm birth.2-Py: The authors declare that they have no competing interests.LM carried out the experiments, analysed the data, and drafted the manuscript. MK & LC designed the Rhea cohort study. HK, MK, EF & LC conceived this metabonomic study, and participated in its design and coordination. EF compiled the epidemiological and clinical data. TA supervised the laboratory work and drafted the manuscript. EH helped with data interpretation and helped to draft the manuscript. MC contributed to the NMR spectral assignment and drafted the manuscript. HK contributed to the statistical analysis and drafted the manuscript. MBT provided her expertise in reproductive epidemiology and statistical analysis. All authors read and approved the final manuscript.Analysis of median differences (Mann-Whitney U test) for all metabolite integrals comparing negative birth outcome groups and controls.Click here for fileDiscrimination between birth outcome cases and control group receiver operating characteristic (ROC) curves for selected candidate metabolites.Click here for file