Recent studies have suggested associations between air pollution and various birth outcomes, but the evidence for preterm birth is mixed.We aimed to assess the relationship between air pollution and preterm birth using 2008–2010 New York City (NYC) birth certificates linked to hospital records.We analyzed 258,294 singleton births with 22–42 completed weeks gestation to nonsmoking mothers. Exposures to ambient fine particles (PM2.5) and nitrogen dioxide (NO2) during the first, second, and cumulative third trimesters within 300 m of maternal address were estimated using data from the NYC Community Air Survey and regulatory monitors. We estimated the odds ratio (OR) of spontaneous preterm (gestation < 37 weeks) births for the first- and second-trimester exposures in a logistic mixed model, and the third-trimester cumulative exposures in a discrete time survival model, adjusting for maternal characteristics and delivery hospital. Spatial and temporal components of estimated exposures were also separately analyzed.PM2.5 was not significantly associated with spontaneous preterm birth. NO2 in the second trimester was negatively associated with spontaneous preterm birth in the adjusted model . Neither pollutant was significantly associated with spontaneous preterm birth based on adjusted models of temporal exposures, whereas the spatial exposures showed significantly reduced odds ratios . Without adjustment for hospital, these negative associations were stronger.Neither PM2.5 nor NO2 was positively associated with spontaneous preterm delivery in NYC. Delivery hospital was an important spatial confounder.Johnson S, Bobb JF, Ito K, Savitz DA, Elston B, Shmool JL, Dominici F, Ross Z, Clougherty JE, Matte T. 2016. Ambient fine particulate matter, nitrogen dioxide, and preterm birth in New York City. Environ Health Perspect 124:1283–1290; http://dx.doi.org/10.1289/ehp.1510266 MaternaPreterm birth is an important predictor of infant mortality, childhood morbidity, and possibly adult morbidity . Analysi2.5 and NO2) and a reduction in term birth weight , New York City Department of Health and Mental Hygiene (NYCDOHMH), to the mothers’ hospitalization records from New York State Department of Health following a study protocol approved by the Institutional Review Board of the BVS NYCDOHMH. Informed consent was not required because this study involved analysis of existing secondary data and posed minimal risk to the subjects. All data were maintained in a password-protected database on a password access–controlled network server at NYCDOHMH with access limited to authorized study investigators. The subset of births used for this analysis was limited to singleton births with 22–42 completed weeks gestation free of congenital malformations to nonsmoking mothers. In addition, we excluded births with an estimated date of conception before 31 July 2007 or after 12 March 2010  to avoid the fixed-cohort bias . Materna2.5 and NO2 at each mother’s residential address at delivery during gestation: NYCCAS measurements to estimate spatial variation of the pollutants across the city and regulatory monitoring data to adjust for the temporal variation. NYCCAS collected 2-week average concentrations of several pollutants at street level in each of the four seasons, December 2008–December 2010. These values were adjusted for week of monitoring and averaged to create annual averages at the monitoring locations; year 1 data were used to fit land-use regression (LUR) models for each pollutant using buffer-based emission and land-use variables and accounting for residual spatial autocorrelation. Annual average pollutant concentration surfaces for NYC were generated from regression model predictions and averaged within 300-m buffers of each maternal address. These maternal address exposure estimates were then temporally adjusted in 2-week intervals using a citywide time series generated from multiple daily regulatory monitors. Year 2 monitoring data were used for validation of the spatiotemporal estimates; predictions were strongly associated with measured concentrations (R2 = 0.83 for PM2.5 and 0.79 for NO2). We averaged over weeks 1–12 and 13–26 to create first- and second-trimester exposure estimates for each birth and calculated third-trimester cumulative exposures by pregnancy week for use in survival analysis  or cesarean section performed (74.x) in the absence of a diagnosis of early or threatened labor  were considered “medically indicated” as were births recorded in the birth records as having labor induction or cesarean section without trial of labor. We adjusted for covariates known to be associated with preterm birth: maternal age, race/ethnicity stratified by birthplace , education , parity , Medicaid status (yes/no), prepregnancy BMI, BMI2, and sex of infant. We adjusted for year of conception to control for long-term trends in pollution levels and birth outcomes. As described in our previous analyses , percent mothers on Medicaid , and percent without a high school diploma , and positively correlated with percent mothers with a college degree  and first births to mothers > 35 years of age . In our study population, hospital-level rates of medically indicated birth (not limited to preterm), after adjusting for demographic factors known to influence the likelihood of medically indicated birth (2.5 and NO2 exposures (Pearson’s ρ ranging from –0.22 to –0.19). Hospitals with higher rates of medically indicated birth showed a distinct spatial pattern ; there was no association between overall average first-, average second-, or cumulative third-trimester PM2.5 exposure and preterm birth , whereas NO2 was not significantly associated with spontaneous preterm birth in either the adjusted or unadjusted models. For the spatial-only exposures, both PM2.5 and NO2 were significantly negatively associated with spontaneous preterm birth, adjusted or unadjusted.Higher overall second-trimester NOrm birth . In the There was no evidence of association between either pollutant and early preterm birth in the adjusted models . Medical2.5 moved slightly above the null , and those for NO2 became slightly more negative . Including ambient temperature in various alternative models had little impact on the estimates; second-trimester PM2.5 OR in the fully adjusted model with temperature was 0.97  compared with 0.99  without, similar to first-trimester NO2 . In the analysis to evaluate alternative spatial random intercepts  moved the apparent negative crude association (model 1) slightly closer to the null. The addition to model 2 of a four-level variable based on quartiles of hospital rates of medically indicated births (model 3) shifted the pollutant effect estimates closer to the unadjusted estimates. The substitution of hospital identifier for census tract in the random intercept  associations between these pollutants and preterm birth in the unadjusted models. Adjusting for individual maternal characteristics, census tract–level social deprivation, random intercepts for either census tracts or hospitals, and the level of medically indicated birth rate at hospitals moved these negative associations for both pollutants closer to the null. The odds ratios for PM2.5 became null, but the associations for NO2 remained negative and statistically significant for the second trimester. By analyzing the spatial-only, temporal-only, and overall exposures, we found that the negative associations were mainly driven by the spatial exposure variation, originating from the common spatial pattern of air pollution and delivery hospitals in which hospitals with higher average exposures among their patients tended to have lower rates of spontaneous preterm birth (a) a negative association between spontaneous preterm birth and air pollution is not biologically plausible, b) the hospital-level rates of medically indicated birth were negatively correlated with maternal air pollution exposures , and c) the pattern of negative associations with the pollutants are very similar for spontaneous preterm births and medically indicated preterm births, the moderate negative association between NO2 and spontaneous preterm birth in the fully adjusted model is a consequence of residual confounding.In this study, we did not find evidence that either PMrm birth . Because2.5 and NO2, fewer effect estimates were available for preterm births studies and that their results were more mixed. Some of the estimates were significantly negative for NO2 . Including a measure of the hospital-level rate of medically indicated labor in our analysis did not attenuate the negative association between air pollution and preterm birth as much as including random intercepts for delivery hospital, suggesting that other provider or patient characteristics that vary among hospitals likely confound the association between air pollution and preterm births, rather than the tendency for medically indicated birth only. To our knowledge, none of the past studies of air pollution and preterm birth have considered potential spatial confounding by hospital of birth.A preterm birth can be spontaneous or medically indicated, and one would expect that the impact of air pollution, if it exists, would be more clearly seen on spontaneous preterm birth than on medically indicated birth, although the censoring of potential future spontaneous preterm births through medical intervention may also affect the observed frequency of spontaneous preterm births. Only two past studies that examined air pollution and preterm birth analyzed spontaneous and medically indicated preterm birth separately: One study , a time-2 and PM2.5 and preterm birth toward the null suggests that refinements on the control of spatial confounding, beyond neighborhood of residence and birth hospital, may help to clarify the underlying nature of associations between these pollutants and preterm birth. Past studies of air pollution and preterm birth, except for 2 in Los Angeles, California, as estimated from temporally adjusted LUR model exposure estimates, but noted that the associations were stronger when estimates were unadjusted for temporal variation. In our analysis, temporal-only exposure estimates showed null associations for PM2.5 and first-trimester NO2, consistent with 2 exposure in our study also is seen in the temporal-only component and deviates from Darrow et al.’s findings. Future studies, especially those using LUR-based exposure estimation, may benefit from analyzing spatial and temporal exposure estimates separately to identify whether the association (or confounding) comes from spatial or temporal components of exposure.Our exposure estimation method was conducive to separately estimating odds ratios for temporal-only and spatial-only exposures. We found that the negative associations were largely driven by the spatial exposure and those associations were more affected by covariate adjustment. Because maternal characteristics of NYC residents are unlikely to have substantively changed in the 3-year study period, the weak influence of covariate adjustment on the temporal-only association was expected. The attenuation of the strong negative association between spatial exposure of NOLimitations of our study include potential exposure misclassification error due to estimates only at mother’s residence at delivery, without details on maternal residential mobility during pregnancy, place and type of work, variation in indoor/outdoor activity, or building characteristics, including floor of residence. The fact that a substantive fraction of the mothers in this study were foreign born leaves some uncertainty regarding the influence of spatial variation of air pollution on the baseline health of these mothers. We were also unable to address all known confounders, including noise and environmental tobacco smoke.2.5 or NO2 and increased numbers of preterm births. The negative association we observed appears to have been confounded by hospital characteristics, which covary with pollutant levels of the patients seen at different hospitals in New York City.Overall, our results do not provide evidence supporting an association between exposure to PM


To investigate the effect of air pollution on birth weight in a medium-sized townin the State of São Paulo, Southeast Brazil. 10, SO2, and O3) and daily averages oftheir concentrations were obtained from the Environmental Sanitation &Technology Company. Statistical analysis was performed by linear and logisticregressions using the Excel and STATA v.7 software programs.  Cross-sectional study using data from live births of mothers residing in São Josédos Campos from 2005 to 2009. Data was obtained from the Department of Informationand Computing of the Brazilian Unified Health System. Air pollutant data. Maternal exposure to PM10 andSO2 during the last month of pregnancy led to lower weight at birth for each 1mg/m Maternal exposure to air pollutants was not associated with low birth weight,with the exception of exposure to SO2 within the last month of pregnancy. This study failed to identify a statistically significant association between thelevels of air pollutants and birth weight, with the exception of exposure toSO The rapid technological advance of the modernworld has resulted in an increase in the quantity and variety of pollutants eliminatedin the atmosphere, affecting the quality of life on the planet.,-,Exposure to air pollutants has shown to be associated with several deleterious effectsto health, even at levels considered safe by environmental legislation.et al in Dominican and African-Americanpregnant women aged 18 to 35 years of age who had lived for at least one year in NewYork, who were nonsmokers without diabetes or hypertension and had negative serology forhuman immunodeficiency virus, indicated that in the population studied, the fetus andthe newborn are more susceptible than adults to toxic environmental substances.A study conducted by Perera Birth weight is an important determinant of neonatal morbimortality and post-neonatalmortality,However, the problem does not occur uniformly among different locations, but rather isrelated to socioeconomic status. The highest percentage of children with LBW isconcentrated in two regions of the world, Asia and Africa, with 27% and 22% of all livebirths showing low birth weight, respectively.,,,LBW has been the subject of several epidemiological studiesThis was a cross-sectional study of data on all births to mothers living in the city ofSão José dos Campos in the years 2005 to 2009, which had completed a live birthcertificate.São José dos Campos is located 80 km from São Paulo, between the Serra do Mar andMantiqueira mountain range, and has a population of approximately 700,000 inhabitants.Its altitude is 600m above sea level and it is located at latitude 23°11' south andlongitude 45°53' west. It has approximately 1,100 industries, especially automotive,pharmaceutical, aerospace, and oil refining. Information on live births was obtained from the Live Birth Information System (SINASC)database through live birth certificates (LBCs), provided by the Department ofInformatics of the Unified Health System (DATASUS), available athttp://www.datasus.gov.br. It included infants born at term, weighing between 1,000g and4,500g, the result of a single pregnancy, with maternal age between 20 and 34 years,after seven or more prenatal consultations and maternal level of schooling of eight ormore complete years. These criteria were adopted to eliminate situations of greatervulnerability, such as preterm newborns and those weighing less than 1,000g, for whichother risk factors might be more relevant than air pollution. 10, SO2, and O3, whichare quantified daily by Companhia de Tecnologia de Saneamento Ambiental (CETESB) of SãoJosé dos Campos. In Brazil, the air quality standards have been established by CONAMAResolution No. 3/1990  and logistic regression(univariate and multivariate) were used. In the linear regression, the variable outcome,birth weight in grams, was analyzed continuously, whereas in the logistic regression,birth weight was dichotomized . ,-,th and 37th weeks of gestation.Regarding maternal exposure to pollutants, the linear regression considered the meanconcentration, in the third trimester and monthly (each month of the third trimester) ofeach pollutant in the respective trimester of gestation, whereas in the logisticregression, these means were recoded into quartiles. The choice of the threemonth-period to estimate maternal exposure to air pollution was based on the fact thatmany studiesThe linear and logistic univariate analyses first assessed the association of birthweight and low birth weight, respectively, with maternal exposure to several pollutants,aiming to estimate the gross effect, i.e., without adjustments, of this exposure on thechild's weight. Furthermore, the univariate logistic model was used to investigate theassociation between outcome and each variable found in the live birth certificates,in order to identify possible confounding factors. In this case, the statisticalanalysis was based on the calculation of the OR to estimate the risk of newborns withlow birth weight associated with each variable. All analyses included 95% confidenceintervals and significance level was set at 5% (α=5%). p-value as the input criteria,provided that this was less than 0.25. If the variables showedp<0.001, the variable entry criteria started from the highest valueof the OR observed in the univariate logistic analysis.Based on the results of the univariate logistic model, variables in the multivariatemodels were selected, in order to control for confounders. First, the multivariate model(linear and logistic) was considered without the pollutant, i.e., the exposure ofinterest. The independent variables were entered one by one in both the linear andlogistic regression, using the smallest p<0.05remained in the final analysis. Only after obtaining the complete model were thepollutants included individually, and their association was tested with birth weight andlow birth weight. Statistical analysis was performed using STATA v.7 . Furthermore, to verify the importance of each variable for the model and its permanencein it, the likelihood ratio was used, and only variables with The study project was approved by the Research Ethics Committee of Universidade deTaubate CEP/UNITAU No. 363/11.Of a total of 45,412 live births in the period of 2005 to 2009, 21,591 births werestudied, considering the inclusion criteria described in the method section. Of theincluded newborns, it was observed that 13,149 (61.2%) were born to mothers who hadpartners, 14,575 (67.5%) were delivered by cesarean section, and 11,012 (51%) weremales . The frep=0.979). 10 andO3 had daily maximum values that were higher than the acceptablevalues, 171 and 209mg/m2 in the last month, indicating that maternal exposureduring this period is associated with a 1.25-fold greater chance of having a newbornwith low birth weight  for the second quartile ofpollutant concentrations. Maternal exposure to each pollutant was analyzed in quartiles and multiple logisticregression was performed using the last three months of pregnancy. In this analysis,adjustment was made for the variables maternal marital status and gender of thenewborn. None of the pollutants showed statistically significant result, with theexception of SO10 and SO2 led to a decreasein mean birth weight: there was a decrease in weight of 3.8g and 38.6g for increasesin mean maternal exposure of 10μg /m10 and SO2, respectively. 3 also appears to indicate a decrease in mean birth weight inthe last trimester of pregnancy and the last two months. However, no result showedstatistical significance. Exposure to Oet alLow birth weight was associated with maternal marital status and gender of the newborn.Despite the predominance of male births, the prevalence of underweight was higher infemale infants. Gouveia et al,The study found a significant association between birth weight and maternal maritalstatus, i.e., mothers living without a partner showed higher risk of having childrenwith low birth weight. These results are consistent with those found by Milleret alet al,Regarding the type of delivery, cesarean birth appears with a lower chance ofinsufficient weight; however, there was no statistically significant result in thepresent study. Antonio 2 in the last month of pregnancy poses an increased risk for low birthweight. The study by Reis2,PM10, and O3 in the second and third trimesters of pregnancy, evenwhen the levels of pollutants found were below the standards established for airquality, i.e., a grade that represents good air quality  in the quartile trimester of pregnancy. There was no statistical significance in the analyses conducted with atmosphericpollutants; however, the multiple logistic analysis showed that maternal exposure toSO quality . Romao e2 and PM10 were not associated withbirth weight, which may result from the low frequency of newborns with low birth weight(3%), adding to that the fact that the mean concentrations of pollutants did not reachvery high values, although the maximum values were inadequate according to air qualitystandards.It can be observed that SORegarding the study limitations, it is worth mentioning that, unlike the exposure datarelated to newborns, maternal factors, prenatal factors, and childbirth, which areobtained per individual, i.e., directly, the data regarding the exposure to airpollutants are obtained by indirect measurement by means of the concentration of airpollutants in the environment, which may hamper the consolidation of more consistentdata. Another limitation refers to obtaining secondary data from DATASUL, an invaluabledata source, but without the direct control of the researcher.2 can lead to low birth weight in the last month of maternalexposure. Although the results were not statistically significant and despite the difficulty toisolate the effect of each pollutant due to the high correlation among them, it wasobserved that SO 10), o ozônio (O3), odióxido de enxofre (SO2), o monóxido de carbono (CO) e os óxidos denitrogênio (NO2).A poluição do ar representa hoje um dos maiores problemas de Saúde Pública, afetandoa saúde dos seres humanos, animais e das plantas. O rápido avanço tecnológico domundo moderno trouxe consigo um aumento na quantidade e na variedade de poluenteseliminados na atmosfera, prejudicando a qualidade de vida em nosso planeta.,-,et al emgestantes dominicanas e afro-americanas entre 18 e 35 anos de idade residentes hápelo menos um ano em Nova York, não fumantes, sem diabetes, hipertensão arterial esoronegativas para o vírus da imunodeficiência humana indicou que, na populaçãoestudada, o feto e o recém-nascido são mais suscetíveis que os adultos às substânciastóxicas ambientais.A exposição aos poluentes do ar vem se mostrando associada a vários efeitosdeletérios sobre a saúde da população, mesmo em níveis considerados seguros pelalegislação ambiental.O peso ao nascer é um importante determinante da morbimortalidade neonatal e damortalidade pós-neonatal,,,,O BPN tem sido alvo de vários estudos epidemiológicosTrata-se de estudo transversal, com dados relativos a todos os nascidos de mãesresidentes no Município de São José dos Campos nos anos de 2005 a 2009 que tiveramsua ficha de declaração de nascido vivo preenchida. São José dos Campos situa-se a 80km de São Paulo, entre as serras do Mar eMantiqueira, e conta com população de aproximados 700 mil habitantes. Sua altitude éde 600m acima do nível do mar e geograficamente está na latitude 23°11' sul e 45° 53'longitude oeste. Conta com cerca de 1.100 estabelecimentos industriais destacando-semontadoras de automóveis, indústrias farmacêutica e aeronáutica e refinaria depetróleo. As informações sobre os nascidos vivos foram obtidas por meio de consulta à base dedados do Sistema de Informações sobre Nascidos Vivos (SINASC) por meio da Declaraçãode Nascido Vivo (DN)¸disponibilizados pelo Departamento de Informática do SistemaÚnico de Saúde (DATASUS), no endereço eletrônico (http://www.datasus.gov.br).Incluíram-se os recém-nascidos à termo, com peso entre 1000g e 4500g, de gestaçãoúnica, com idade materna entre 20 e 34 anos, sete ou mais consultas de pré-natal eescolaridade materna de 8 ou mais anos de estudo concluído. Esses critérios foramadotados para se eliminar situações de maior vulnerabilidade como o recém-nascidopré-termo e o peso abaixo de 1.000g, para os quais outros fatores de risco poderiamser mais relevantes do que a poluição atmosférica.10, SO2 e O3, quesão quantificados diariamente pela estação medidora da Companhia de Tecnologia deSaneamento Ambiental (CETESB) de São José dos Campos. No Brasil os padrões dequalidade do ar foram estabelecidos pela Resolução CONAMA nº 3/1990Os poluentes estudados foram o PM,-,a e a 37a semana de gestação.Com relação à exposição materna aos poluentes, na regressão linear foram consideradasa média do 3º trimestre e mensal (dos meses do 3º trimestre) da concentração de cadapoluente no respectivo trimestre de gestação, enquanto, na regressão logística, taismédias foram recodificadas em quartis. A escolha do período de três meses para aestimativa da exposição da mãe à poluição do ar se baseou no fato de que muitosestudos,As análises univariadas linear e logística examinaram, primeiramente, a relação dopeso e do baixo peso ao nascer, respectivamente, com a exposição materna aos diversospoluentes com o objetivo de estimar o efeito bruto, ou seja, sem ajustes, dessaexposição no peso da criança. Além disso, o modelo logístico univariado foi utilizadopara verificar a relação do desfecho com cada variável presente na Declaração deNascido Vivo , de maneira a identificar possíveis fatores de confusão. Neste caso, aanálise estatística baseou-se no cálculo do OR para se estimar o risco derecém-nascido com baixo peso ao nascer associado a cada variável. Em todas asanálises foram construídos intervalos de 95% de confiança e adotado nível designificância de 5% (α=5%).p<0,001, o critério daentrada das variáveis partiu do maior valor do OR observado na análise logísticaunivariada. E ainda, para verificação da importância de cada variável para o modelo esua permanência neste, utilizou-se a razão de verossimilhança (likelihood-ratiotest), permanecendo no final da análise apenas as variáveis comp<0,05. Apenas após obter o modelo completo, os poluentesforam incluídos individualmente e testou-se sua associação com o peso ao nascimento eo baixo peso ao nascer. A análise estatística foi realizada no STATA v.7.  A partir dos resultados deste modelo logístico univariado, foram selecionadas asvariáveis para os modelos multivariados, com a finalidade de se fazer o controle deconfusão. Primeiramente o modelo multivariado (linear e logístico) foi consideradosem o poluente, ou seja, a exposição de interesse. As variáveis independentesentraram uma a uma, tanto na regressão linear quanto na logística, utilizando-se comocritério da ordem de entrada o menor valor do "p", contanto que este fosse menor que0,25. Caso as variáveis apresentassem o 363/11.O projeto deste estudo foi avaliado e aprovado pelo Comitê de Ética da Universidadede Taubaté CEP/UNITAU nDo total de 45.412 nascidos vivos do período entre 2005 e 2009, foram estudados21.591 nascimentos, considerando-se os critérios de inclusão apresentados nametodologia. Destes incluídos, observou-se que 13.149  eram filhos de mães quepossuíam parceiro, 14.575  nasceram de parto tipo cesáreo e 11.012 (51%) eramdo sexo masculino . A frequp=0,979). A 10 e O3tiveram seus valores máximos diários maiores do que os valores aceitáveis,respectivamente, 171 e 209 µg/mA 2 no último mês, indicando que a exposição materna nesse período seassocia à 1,25 vezes mais chance de recém-nascido de baixo peso  para o 2º quartil da concentração do poluente.A exposição materna a cada poluente foi analisada em quartis e realizada regressãologística múltipla segundo os últimos três meses de gestação. Nesta análise, houve oajuste para as seguintes variáveis: estado civil materno e sexo do recém-nascido.Nenhum poluente apresentou resultado estatisticamente significante, com exceção doSOA 10 e ao SO2 levou a umadiminuição da média do peso ao nascer: houve decréscimo do peso de 3,8g e 38,6gparaum aumento na exposição média materna de, respectivamente, 10μg/m10 e do SO2. Aexposição ao O3 também parece indicar essa diminuição da média do peso aonascer no último bimestre e trimestre de gestação. No entanto, nenhum resultadoamostrou significância estatística.Já a et alO baixo peso ao nascer esteve associado ao estado civil materno e ao sexo dorecém-nascido. Apesar do predomínio de nascimentos do sexo masculino, a prevalênciade baixo peso foi maior nas crianças do sexo feminino. Gouveia et al,No estudo encontrou-se associação significativa com o peso ao nascer e o estado civilmaterno, ou seja, mães vivendo sem o parceiro mostram risco maior de terem filhos combaixo peso ao nascer. Estes resultados são consistentes com os encontrados porMonteiro et alet al,Com relação ao tipo de parto, o cesáreo aparece com menor chance de ocorrência depeso insuficiente; no entanto, não se encontrou resultado estatisticamentesignificante no presente estudo. Antonio 2 durante o último mês gestacional apresenta um maior riscopara o nascimento de uma criança com baixo peso. No estudo de Reis,2, PM10 eO3 no 2º e 3º trimestres de gestação, mesmo quando os níveis dasconcentrações dos poluentes se encontravam abaixo dos padrões estabelecidos para aqualidade do ar, ou seja, com "nota" que representa boa qualidade do ar  no 3º trimestre de gestação.Não se encontrou significância estatística nas análises realizadas com os poluentesatmosféricos, porém, foi observado na análise logística múltipla que a exposiçãomaterna ao SOde do ar . Romão e2 e PM10 não mostraram associação como peso do RN, o que pode decorrer da baixa frequência de recém-nascidos com baixopeso ao nascer (3%), somando-se a isso as médias das concentrações dos poluentes nãoatingiram valores muito altos, apesar dos valores máximos serem inadequados segundoos padrões da qualidade do ar.Pôde-se observar que o SOCom relação às limitações deste estudo, vale comentar que, diferentemente dos dadosde exposição relativos ao recém-nascido, à mãe, ao parto e pré-natal, que se obtémindivíduo a indivíduo, ou seja, de forma direta, os dados referentes à exposição aospoluentes do ar são obtidos por medida indireta, por meio da concentração dospoluentes do ar no ambiente, o que pode dificultar a consolidação de dados maisconsistentes como os individuais. Outra limitação a ser apontada refere-se à obtençãode dados secundários do DATASUS, fonte valiosa de dados, porém sem o controle diretodo pesquisador como se tem com a coleta de dados de fontes primárias.2 pode levar ao baixo peso ao nascer, no último mês de exposiçãomaterna. Apesar dos resultados não serem estatisticamente significantes e da dificuldade emisolar o efeito de cada poluente, devido à alta correlação entre eles, observou-seque o SO


Birth weight of a child is an important indicator of its vulnerability for childhood illness and chances of survival. A large number of infant deaths can be averted by appropriate management of low birth weight babies and prevention of factors associated with low birth weight. The prevalence of low birth weight babies in Nepal is estimated to be about 12-32%.Our study aimed at identifying major determinants of low birth weight among term babies in Nepal. A hospital-based retrospective case control study was conducted in maternity ward of Tribhuvan University Teaching Hospital from February to July 2011. A total of 155 LBW babies and 310 controls were included in the study. Mothers admitted to maternity ward during the study period were interviewed, medical records were assessed and anthropometric measurements were done. Risk factors, broadly classified into proximal and distal factors, were assessed for any association with birth of low-birth weight babies. Regression analysis revealed that a history of premature delivery , hard physical work during pregnancy , younger age of mother , mothers with haemoglobin level less than 11gm/dl  and lack of consumption of nutritious food during pregnancy  were significantly associated with the birth of LBW babies. These factors should be addressed with appropriate measures so as to decrease the prevalence of low birth weight among term babies in Nepal. Birth weight is an important determinant of an infant’s survival and future development . Low birPrior studies from the Indian subcontinent have identified several socio-economic factors in addition to gestational age and maternal health as risk factors for SGA babies in this region –18. HoweA retrospective case-control study design was used. The study was conducted in Tribhuvan University Teaching Hospital (TUTH), one of the largest tertiary hospitals in Kathmandu, Nepal. Cases were identified as newborn babies with birth weight less than 2500 grams whereas controls were newborn babies with a birth weight of more than or equal to 2500 grams. Our study population included singleton newborn babies at term without any congenital diseases. The study was limited to SGA babies at term only. Unlike in developed countries, majority of LBW babies in the developing countries are small for gestational age due to intrauterine growth retardation, which has been associated with a variety of socio-economic and maternal risk factors ,12,16.A total of 155 eligible consecutive cases of LBW babies at term were admitted to TUTH between February 2011 and July 2011. Age and sex matching with two controls was done on the same day when a case was found. Whenever there were more than 2 eligible controls for a case, controls were randomly selected. A total of 310 controls were selected. The non-selected controls were included for other cases on the same day if it met the inclusion criteria (i.e. age and sex matched). Due to high volume of deliveries, controls for a case were found on the same day. The number of cases and controls was calculated by taking power at 80%, odds ratio of 1.8, two sided significant level at 0.05 and proportion of controls with exposure as 0.25 using Epi Info (Version 3.5.2).The data was collected by interviewing the mothers, observing medical records of the newborn and measuring their anthropometry. Birth weight was measured by the study investigators using a digital non-hanging type salter scale and rounded to the nearest 10 grams. The instrument was calibrated each time before use with a standard 1 kilogram weight. Data were collected by medical and public health students who were properly oriented on data collection techniques by the principle investigator. A structured questionnaire was adapted from previous study by WHO in South-East Asia region and other studies conducted in this region. The tools were pretested during the month of January in Tribhuvan University Teaching Hospital itself in five percent of the sample. The ratio of cases and controls in pretesting was 1:1. Necessary amendments were made and measuring tools were checked properly prior to data collection. Thus collected data was reviewed, edited and re-coded. The study was ethically approved by Institutional Ethical Review Board of Maharajgunj Medical Campus, Institute of Medicine. Respondents were clearly explained about the objective and voluntary nature of the study. Written consent was taken from the respondents and their guardians prior to the interview. Participants were explained about their right to refuse participation and withdraw from the study.The explanatory variables were divided into proximate and distal factors. The proximate factors included history of chronic medical illness, history of abortion, history of premature delivery, hard physical work during pregnancy, illness during current pregnancy, age of mother, parity, height of mother, haemoglobin level, the practices of taking iron tablets, consuming nutritious food  daily, drinking alcohol during pregnancy and antenatal care (ANC) visits during last pregnancy. Chronic medical illness was defined as a pre-existing medical illness of the mother that was documented in the medical record of TUTH with an onset prior to the current pregnancy. Regular physical work during pregnancy was identified as moderate or hard. Hard physical work comprised of three or more of the following listed works: daily household chores, fetching water with large buckets, lifting heavy loads, chopping woods, cutting grass for cattle feeding and washing clothes/utensils for long. These works have been accounted under hard physical activity in a study among rural women in India . On the Participants were categorized into two groups based on haemoglobin level; one with haemoglobin level less than 11gm/dL and the other with haemoglobin level 11gm/dL or more based on a prior study . The iroThe distal factors included education and occupation status of the parents, ethnicity, place of residence, type of family and income of the family. Parental educational status was grouped as illiterate, primary, lower-secondary, secondary, intermediate, bachelors and above. They were further categorized into primary level or less and secondary or above. Occupations of the parents were categorized into five groups; agriculture, service, business, household works and others. These occupations are grouped into either labor intensive or non-labor intensive groups. Labour intensive occupation included more physical works under lower positions and with comparatively less wages. Ethnicity of the mother was first categorized into six major ethnic groups  as per the classification of the Health Management Information System of Ministry of Health/Nepal . In the The data was entered in Epi info (Version 3.5.2)and analyzed in Statistical Package for Social Sciences (Version 17). Bivariate analysis was first conducted with chi-squared test to assess the differences of the various proximal and distal factors between the cases versus the controls see . All varA total of 155 cases and 310 controls were studied. Tables Chronic medical illness was present only in a small percentage of the study population. Majority of the mothers had no prior history of abortion or premature delivery. Engaging in hard physical work during pregnancy was fairly common (46% and 33% among cases and controls respectively). Majority of mothers belonged to age 20 to 29 years (69% among cases and 75% among controls), had a height of 146 cm or more (89% among cases and 94% among controls). More than 90% of both groups were taking iron supplements since at least 3 months. Majority of mothers (93% cases and 87% controls) had their haemoglobin level less than 11gm/dl. Alcohol intake during pregnancy was fairly low (0.6% and 1.3% respectively) and majority of mothers had an ANC visit of at least 4 times (96.8% and 93.5% respectively) as in Results of the bivariate Tables  and 2 anBased on the findings of the present study, history of premature delivery, hard physical work during pregnancy, earlier age at pregnancy, lack of consumption of nutritious food during pregnancy and maternal anemia were the independent risk factors for the birth of SGA babies in Nepal.Our study revealed that a history of premature delivery was significantly associated with LBW in babies. A study by Singh et al in Ahemdabad, India also showed similar findings . A highlHard physical work during pregnancy was found to be significantly associated with LBW in this study. A greater proportion of LBW babies were born to mothers who had done hard physical work during pregnancy. A study from Lithuania depicted similar findings that hard manual work by mothers attributed to greater percent of LBW cases . AgarwalMany studies have shown age of the mother to be significantly associated with the birth of LBW babies ,36–41. PThe current study well found mother’s level of haemoglobin to be associated with LBW. Lower concentration of maternal haemoglobin is one of the risk factors for LBW among children ,44. MateThe study also found that consumption of nutritious diet during pregnancy affects the birth weight, a well-established fact –54. PoorThis study however had some notable limitations. The findings might be influenced by purposive selection of study area, study design bias and social-desirability bias. Use of hospital records for background information may have introduced some bias. The study was based on a single tertiary institution limiting its generalizability. In a country with proportion of home deliveries about 65% , a commuHistory of premature delivery, hard physical work during pregnancy, younger age at pregnancy, mother’s lower haemoglobin level and lack of nutritious diet consumption during pregnancy were the major determinants of low birth weight among term babies in Nepal. Public health programs should focus on raising awareness on avoiding early marriage and pregnancy of females. Moreover, it should provide emphasis on adequate rest and nutrition during pregnancy in order to decrease the prevalence of low birth weight. Provision of a more intensive ANC to mothers with a history of premature deliveries can be another important strategy to prevent low birth weight babies. The role of family members is important especially in fulfilling the nutritional and health care needs of the pregnant mothers along with supporting her to take adequate rest in the cultural context of Nepal.


We determined the association of cord blood 25-hydroxyvitamin D [25(OH)D] with birth weight and the risk of small for gestational age (SGA). As part of the China-Anhui Birth Cohort (C-ABC) study, we measured cord blood levels of 25(OH)D in 1491 neonates in Hefei, China. The data on maternal sociodemographic characteristics, health status, lifestyle, birth outcomes were prospectively collected. Multiple regression models were used to estimate the association of 25(OH)D levels with birth weight and the risk of SGA. Compared with neonates in the lowest decile of cord blood 25(OH)D levels, neonates in four deciles  had significantly increased birth weight and decreased risk of SGA. Multiple linear regression models showed that per 10 nmol/L increase in cord blood 25(OH)D, birth weight increased by 61.0 g  at concentrations less than 40 nmol/L, and then decreased by 68.5 g  at concentrations from 40 to 70 nmol/L. This study provides the first epidemiological evidence that there was an inverted U shaped relationship between neonatal vitamin D status and fetal growth, and the risk of SGA reduced at moderate concentration. Growth restriction in utero increases infant morbidity and mortality as well as the risk of adult chronic diseases1245Maternal vitamin D deficiency during pregnancy is a significant public health problem globally79101112131416171819Most studies assessed the associations of maternal 25-hydroxyvitamin D (25[OHD]) in early pregnancy with fetal growth. Evidences from some studies support the hypothesis that lower vitamin D status causes adverse fetal growth in a linear way, more severe deficiency would be expected to have a stronger effect than less severe deficiency or sufficiency91012131415181921Notably, increased risks for a range of health outcomes, including SGA, among pregnant women with high 25(OH)D levels has been reported by several studies1118P < 0.05 for the trend). Male infants had slightly higher cord blood 25(OH)D levels than females, but the difference did not reach statistical significance after adjusting for the season of birth (P = 0.11).Attrition analyses showed the distributions of sociodemographic characteristics, health status, lifestyle and birth outcomes, non-participants did not differ from participants. In this study, the mean age of the participants was 27.65 years (SD = 3.66), the majority (79.2%) of participants had completed high school and beyond (>9 years), and the income of 84.9% of the participants was more than 2 000 RMB per month. The mean gestational age at delivery was 38.91 weeks (SD = 1.47), ranging from 32 to 42 weeks, with 96 preterm births (5.3%). The mean birth weight was 3383.1 g (SD = 453.9), ranging from 1500 to 5500 g, with 133 small for gestational age infants (8.9%). In our sample, 791 (53.1%) newborns were males. The mean cord blood 25(OH)D level was 39.43 nmol/L (SD = 20.35), ranged from 6.06 to 119.64. In a multiple linear regression model, the deciles of cord blood 25(OH)D were significantly associated with birth weight, even after adjustment for gestational week, infant gender, prepregnancy BMI, gestational weight gain (GWG), birth season, maternal age, education, income, alcohol consumption, pregnancy complications, parity, paternal smoking and alcohol consumption. Neonates in four deciles  had significantly increased birth weight compared with those in the lowest decile. The highest adjusted regression coefficients was observed in the fifth decile  .Consistently, in a multiple logistic regression model with full adjustment described above , the deciles of cord blood 25(OH)D were significantly associated with risk of SGA. Neonates in four deciles  had significantly decreased risk of SGA compared with those in the lowest decile, with the lowest adjusted OR observed in the fifth decile [0.34 (95% CI 0.14–0.80)] . Figure P < 0.001). Indeed, this model suggested a significant steep increase in birth weight up to 40 nmol/L of cord blood 25(OH)D and a gradual downward trend in birth weight at concentrations from 40 to 70 nmol/L, and then level off thereafter (We further assessed the dose-response relation between cord blood 25(OH)D levels and birth weight. The inverted U-shaped nature of the association of cord blood 25(OH) D with birth weight was confirmed with Gaussian curve-fitting model (ereafter .Multiple linear regression model shows that per 10 nmol/L increase in cord blood 25(OH)D levels, birth weight increased by 61.0 g  at concentrations less than 40 nmol/L, but decreased by 68.5 g  at concentrations from 40 to 70 nmol/L. We further assessed the difference of association between cord blood 25(OH)D and birth weight in male and female infants. After adjustment for confounders, average birth weight of boys and girls increased by 62.6 g  and 68.0 g , respectively, per 10 nmol/L increase in cord blood 25(OH)D up to 40 nmol/L. However, average birth weight of boys and girls decreased by 58.7 g  and 82.3 g , respectively, per 10 nmol/L increase in cord blood 25(OH)D from 40 to 70 nmol/L .To our knowledge, this is a study with the largest sample examining the association of cord blood 25(OH)D concentrations with birth weight and the risk of fetal growth restriction. The estimated difference in mean of birth weight among neonates in the fifth decile compared with those in the lowest decile was around 176 g, greater than the reduction in birth weight observed for children exposed to maternal smoking in uteroMost previous studies in this field were undertaken in developed countries131413262728Currently, there is limited information on the association between fetal vitamin D status and fetal growth. A small cross-sectional study in China demonstrated that average birth weight was lower in newborns (n = 27) with 25(OH)D < 25 nmol/L than those (n = 31) with 25(OH)D ≥25 nmol/LPrevious studies have several limitations including small sample sizes, inconsistent definitions for fetal vitamin D status, few adjustment for confounders and only linear models performed. In this study with a large sample, adjustment for more confounders and using non-linear models, we found that there was an inverted U-shaped relationship between cord blood 25(OH)D and birth weight, suggesting that the newborn with low or high levels of fetal vitamin D are both in increased risk of growth restriction and higher levels of fetal vitamin D have not been shown to confer greater benefits. However, it should be emphasized that the inverted U shaped relationship in this study does not imply causality, because fetal weight would have to have been affected by 25(OH)D at critical times of development rather than at birth when samples were collected. Additionally, although there was an inverted U shaped relationship, the infants above the seventh decile of 25(OH)D levels still had higher birth weight than those in the first decile.The finding that vitamin D levels in the upper range increased the health risk in this study was consistent with previous studies examining the risk of fetal growth restriction30343638Our study has three major strengths. First, it is the first study reporting that there was an inverted U-shaped relationship between cord blood 25(OH)D and fetal growth. Second, prospective data collection procedure, a relatively large sample size and statistical adjustment for a large number of confounders were used. Third, utility of both continuous measure and the deciles of cord blood 25(OH)D concentrations helped to detect the inverted U-shaped relation.However, several limitations in this study should also be acknowledged. First, we only measured 25(OH)D concentrations, which is one member of the vitamin D family. At birth, there are potential changes to levels of vitamin D binding protein, 1,25(OH)2D, the CYP27 enzyme and expression of vitamin D receptorsIn conclusion, this study provides the first epidemiological evidence that there was an inverted U shaped relationship between neonatal vitamin D status and fetal growth. The findings suggested that vitamin D supplements during pregnancy should be encouraged, but with caution. If confirmed in further studies, it could have important implications for public health policy.This study was conducted on the basis of a prospective birth cohort comprising 2552 pregnant women recruited in Hefei (32 °N latitude) from January to September 2008. As part of the C-ABC studyCord blood sample was collected immediately after delivery and anticoagulated by use of sodium heparin. Plasma samples were centrifuged and promptly refrigerated at −4 °C, within 12 h, transferred to −80 °C freezers for long-term storage. 25(OH)D concentrations were measured using the commercial radioimmunoassay kits . Intra-assay and inter-assay coefficients of variation were 8.8% and 11.1% respectively. Plasma concentrations of 25(OH)D were analyzed both as a continuous variable and a decile.The key outcomes assessed in this study included birth weight and SGA defined as birth weight <10th percentiles of distribution for gestational age and infant genderThe most important potential confounder was gestational age, infants gender, prepregnancy body mass index (BMI), GWG and birth season, which have been known to be associated with birth weight or maternal 25(OH)D levels15Cord blood 25(OH)D concentrations were divided into deciles. General linear regression models and Mantel-Haenszel chi-square test were used to compare means and proportions for the characteristics of pregnancy women across the deciles of 25(OH)D level. The means of the deciles were fit as continuous variables to estimate the trend of variables across deciles in a linear regression model. Differences of birth weight according to maternal characteristics were evaluated using linear regression models.The adjusted regression coefficients of multiple linear regression models were generated for the association between the deciles of 25(OH)D levels and birth weight, that were adjusted for gestational week, infant gender, prepregnancy BMI, GWG and birth season, and further adjusted for maternal age, education, income, alcohol consumption, pregnancy complications, parity, paternal smoking and alcohol consumption. Multivariable logistic regression models were used to determine odd ratios (OR) and 95% CIs for SGA after adjusting for confounders as described above.The curvilinear association between 25(OH)D and birth weight was best described using a Gaussian curve-fitting modelHow to cite this article: Zhu, P. et al. Cord Blood 25-hydroxyvitamin D and Fetal Growth in the China-Anhui Birth Cohort Study. Sci. Rep.5, 14930; doi: 10.1038/srep14930 (2015).


This study investigated whether the glycoside asiaticoside inhibits Aβ1–42 fibrillation steps, more Aβs would remain free and rapidly diffuse in the confocal volume. In contrast, “weaker or no inhibition” permits a greater number of Aβs to polymerize into oligomers, leading to fibers and gives rise to slow diffusion times in the solution. Trace amounts of 5-carboxytetramethylrhodamine (TAMRA)-labeled Aβ1–42 in the presence of excess unlabeled Aβ1–42 (10 μM) was used as a fluorescent probe. Steady-state and kinetic-Thioflavin T (ThT) fluorospectroscopy, laser-scanning fluorescence microscopy (LSM), and transmission electron microscopy (TEM) were also used to monitor fibrillation. Binding of asiaticoside with Aβ1–42 at the atomic level was computationally examined using the Molegro Virtual Docker and PatchDock.Fluorescence correlation spectroscopy (FCS), evaluating the Brownian diffusion times of moving particles in a small confocal volume at the single-molecule level, was used. If asiaticoside inhibits early Aβ1–42 was 208 ± 4 μs, which decreased to 164 ± 8.0 μs in the presence of asiaticoside, clearly indicating that asiaticoside inhibited the early stages Aβ1–42 of fibrillation, leaving more free Aβs in the solution and permitting their rapid diffusion in the confocal volume. The inhibitory effects were also evidenced by reduced fiber formation as assessed by steady-state and kinetic ThT fluorospectroscopy, LSM, and TEM. Asiaticoside elongated the lag phase of Aβ1–42 fibrillation, indicating the formation of smaller amyloid species were impaired in the presence of asiaticoside. Molecular docking revealed that asiaticoside binds with amyloid intra- and inter-molecular amino acid residues, which are responsible for β-sheet formation and longitudinal extension of fibrils.With 1 h of incubation time for aggregation, FCS data analysis revealed that the diffusion time of TAMRA-AβFinally, asiaticoside prevents amyloidogenesis that precedes neurodegeneration in patients with Alzheimer’s disease. Lag time (LT) was defined as the time at which the tangent at the point of the maximum fibrillation rate intersects the abscissa. At 95% confidence interval, T50  was increased by asiaticoside  . TAMRA-Aβ1–42 evoked characteristic orange fluorescence . Again, the number of red spot areas was significantly lesser in the presence of asiaticoside, indicating asiaticoside inhibited the fibrillation and consequently the amyloid deposits. Total areas of red-fluorescence areas were, respectively, 2513 ± 9.4 and 481 ± 18.7 μM2 for the control and asiaticoside-treated samples.The fibrillations of unlabeled Aβ) Figure A, B, whie Figure C, D. Thes Figure A, B, sugs Figure C, D are 1–42 alone) contained abundantly aggregated amyloid fibrils in the assembly buffer, whereas the Aβ1–42 + asiaticoside samples contained only very small amounts of aggregates. The former fibrils exhibited a typical filamentous and branching morphology  identified amino acids 1–18 that remained in a generally disordered region. These amino acids do not adopt a stable structure and cannot engage in sufficient numbers of favorable intrachain interactions  and 31–42 (β2). Intermolecular side chain contacts are formed between the odd-numbered residues of strand β1 of the nth molecule and the even-numbered residues of strand β2 of the (n − 1)th molecule. This interaction pattern leads to partially unpaired β-strands at the fibrillar ends, which explains the sequence selectivity, cooperativity, and apparent unidirectionality of Aβ fibril growth [The ANCHOR server . The score  mimics the potential energy change, when the protein and ligand come together. This means that a very negative score corresponds to a strong binding and a less negative or even positive score corresponds to a weak or non-existing binding. Water molecules were not included in the docking setup, thus the ‘water-ligand interaction’ did not contribute to the MolDock score. The MolDock score of pose 1 was lower than that of the other four poses , indicating the structure in pose 1 was superior. The MolDock scores of the five poses for docking with the dimer were as follows: pose 1 = −134.48, pose 2 = −118.74, pose 3 = −118.64, pose 4 = −113.89, and pose 5 = −109.09 Kcal/mol. The results of the computational docking of asiaticoside with the A1monomer and A1-B1dimer are shown in Figure 1–42 dimer was docked with asiaticoside, the amino acid residues Leu17, Val18, Phe19, Val36, Val40, Ile41 and Ala42 of A1-monomer, and those of the Val40 and Ala42 of B1-monomer bound with asiaticoside  was visualized by Marvin Space Figure E, 9G. Af1–42 fibrillation in FCS. The initial steps of Aβ aggregation are preceded by the transformation of α-sheets to β-sheets, through lateral stacking of dimers, trimers, tetramers, and oligomers, leading to matured amyloid fibers [1–42 alone. The increased diffusion time of the latter is consistent with its increased molecular mass. At time zero, the diffusion time of TAMRA-Aβ1–42 was not significantly different in the presence or absence of asiaticoside, demonstrating that the amyloids were in their monomeric state at the start and/or early steps of fibrillation. After 1 h of incubation, the diffusion time of TAMRA-Aβ1–42 in the TAMRA-Aβ1–42 + unlabeled Aβ1–42 samples increased to 208 ± 4 μs, indicating that amyloid species/units with a molecular mass relatively greater than the previous one occurred in the early steps, including oligomrizations of Aβ fibrillation process. This resulted in a concurrent slowness of movements of the amyloid units during “entering-and-leaving” the confocal volume of FCS. In FCS, larger amyloid aggregates (of Aβ1–42) in solution diffuse less frequently through the confocal volume, thereby resulting in longer diffusion times. Thus, the decreased diffusion time in the presence of asiaticoside reflects an inhibition of Aβ1–42 aggregation by this glycoside, allowing more Aβ1–42 in the confocal volume to freely diffuse in and out of the confocal volume. Repeated FCS measurements of 10-s sampling times illustrated an increase in the amplitude (y-axis). The amplitude reflects the inverse of the average number of fluorescent TAMRA-Aβ1–42 molecules in the focal volume. In other words, the correlation amplitude increases in size as the intensity fluctuation grows larger. This again demonstrates that asiaticoside inhibited amyloid fibrillation, allowing an increasing number of free TAMRA-Aβ1–42 molecules to freely diffuse in the confocal volume. The diffusion time decreased as the asiaticoside concentration increased  is a method based on fluorescence intensity fluctuations that result from dynamic movement and fluorescence quantum yield of fluorescent molecule while it diffuses in and out of small detection area. In this study we recorded the effect of asiaticoside on the dynamical properties of Aβd fibers . Autocord Figure C.C. asiatica extract, which contains two major triterpene glycosides (asiaticoside and madecassoside) as active components, is increasingly being used to enhance memory [1–42 fibrillation [1–42 fibrillation at the early stages of the process, hinting at the basis of the neuroprotective effects of C. asiatica extract and/or its active components (asiaticoside and/or madecassoside) against Aβ1–42–induced AD.Recently, e memory , which iillation . In the 1–42 fibrillation. The ThT growth curve for Aβ1–42 was sigmoidal showing a dependence on time, which suggests that the amyloid assembly process is cooperative. The presence of asiaticoside led to a significant increase in the lag phase of Aβ1–42 fibrillation with a concurrent deceleration of the exponential growth of fibrils. Thus it would appear that the asiaticoside acted to reduce the rate of nucleation, and, however, once critical nuclei were formed elongation was also affected by the presence of the asiaticoside. Once the FCS measurements and kinetic and/or steady-state ThT-fluorospectroscopic measurements proved that asiaticoside delayed/inhibited fibrillation, we also examined the effect of asiaticoside on amyloid fibrillation using Laser Scanning Microscopy (LSM). The ThT-bound amyloidal aggregates (green) were clearly visible in the confocal images. The red fluorescence areas indicate the larger aggregates of TAMRA-Aβ1–42-intercalated fibers of Aβ1–42 (5 nM TAMRA-Aβ1–42 was used in excess of unlabeled Aβ1–42). Total numbers of fluorescent aggregates of Aβ1–42 control samples were much more frequent and larger than those of the Aβ1–42 + asiaticoside samples  at which asiaticoside exerted the inhibitory effect on fibrillation. Therefore, we evaluated the effect of asiaticoside on the kinetics of Aβ1–42 in the grid fields of TEM (Figure 1–42 goes through a diverse dynamic exchange of conformational stages and includes continuous shuffling and growth, particularly, during the first stages of aggregation; Aβ1–42 can adopt, at least, a globular structure or so-called nucleating center. These globular structures form oligomers leading to protofibrils, which then associate in elongated fibers, finally to the highest ordered matured Aβ1–42 fibers. Asiaticoside-induced interference of Aβ1–42 fibrillation also was evident from the morphology of Aβ1–42 fibrils in transmission electron micrographs. Consistent with the fact that pretreatment of Aβ1–42 with asiaticoside kinetically lengthened the formation of amyloid nucleating centers and reduced the protofibrillar structures, TEM grids of the Aβ1–42 + asiaticoside samples were also devoid of complete fibers, and only a small number of fibrils exhibited a diffused, shorter and some spheroid-type morphology. Structural characterization of these spheroidal species has remained elusive. Soluble amyloid β-oligomeric intermediates, emerging during amyloid fibril assembly, represent the main molecular species responsible for toxicity to cells and tissues [1–42 into unstructured and nontoxic off-pathway oligomers. Asiaticoside-induced decreases in the formation and quantity of amyloid fibers (Figures 1–42 samples are not toxic oligomer species. Otherwise, asiaticoside containing centella asistica extract couldn’t have beneficial effects on memory [1–42-induced neurotoxicity in AD model animals [1–42 may be mediated by the effects of this glycoside.We directly observed the high-resolution image of the dense aggregates of matured fibers, long/short or single/few scattered matured fibrils, typical branched fibrils, protofibrils, oligomer-like, in some instances, amorphous and other unstructured smaller aggregates of AβM Figure . This su tissues . The pre tissues . Ehrnhoe tissues  showed t Figures  and 6 arn memory , dementin memory , neuronan memory , neuriten memory  and Aβ1– animals ,41. Our  animals . Because1–42 monomers and dimer was used in combination with experimental data to obtain information on the binding domain and mechanism of inhibition. The ability of Aβ1–42 to bind to itself and to different ligands in a highly specific manner is an important feature of amyloidogenesis. Kirschner et al. [1–42, were located in the regions of residues 17–22 and 32–42 Figure 1–42 also revealed that asiaticoside bound with the amino acid sites located at the β1 and β2 domain of the Aβ1–42 (Figure 1–42 protofibrils, leading to matured fibers [Structure-based drug design has made tremendous contributions to the fields of cancer chemotherapy and drug-resistant infections. Computational structure-based drug design may, therefore, facilitate the development of novel treatments for AD. Therefore, the docking of asiaticoside with the Aβr et al. . The agg2 Figure E-H. Rema2 Figure . These rd fibers . This st1–42 and interactions between small molecules, such as, asiaticoside and Aβ1–42 may contribute to the development of inhibitors against Alzheimer’s disease. In this study, by means of a combined approach of ThT, FCS, LSM, TEM techniques, we suggest a model of how the asiaticoside interacts with β1–42. Our experimental results allow us to conclude that asiaticoside inhibits early stages of fibrillogenesis through interactions with ‘nucleating’ amyloid species and decelerating the growth phase. In addition, the in silico, docking of asiaticoside with Aβ1–42 appears to be consistent with the ability of this glycoside to inhibit fibrillation by acting both at the intra- and inter- β-sheet interaction sites. Finally, the results obtained support further development of asiaticoside for clinical use in Aβ1–42 -induced neurodegenerative diseases, such as Alzheimer’s disease.Studies of aggregation of Aβ


Infant mortality reflects not only the health of infants but societal well-being as a whole. This study explores distal socioeconomic and related proximate determinants of infant mortality and provides evidence for designing targeted interventions.Survival information on 5391 live born infants (2006–2010) was examined from the nationally representative Nepal Demographic Health Survey 2011. Bivariate logistic regression and multivariate hierarchical logistic regression approaches were performed to analyze the distal-socioeconomic and related proximate determinants of infant mortality.Socio-economic distal determinants are important predictors for infant mortality. For example, in reference to infants of the richest class, the adjusted odds ratio of infant mortality was 1.66 (95 % CI: 1.00–2.74) in middle class and 1.87 (95 % CI: 1.14–3.08) in poorer class, respectively. Similarly, the populations of the Mountain ecological region had a higher odds ratio  of experiencing infant mortality compared with the populations of the Terai plain region. Likewise, the population of Far-western development region had a higher adjusted odds ratio  of experiencing infant mortality than the Western development region. Moreover, the association of proximate determinants with infant mortality was statistically significant. For example, in reference to size at birth, adjusted odds ratio of infant dying was higher for infants whose birth size, as reported by mothers, was very small  than whose birth size was average. Similarly, fourth or higher birth rank infants with a short preceding birth interval  were at greater risk of dying  compared to the second or third rank infants with longer birth intervals. A short birth interval of the second or the third rank infants also increased the odds of infant death .Socioeconomic distal and proximate determinants are associated with infant mortality in Nepal. Infant mortality was higher in the poor and middle classes than the wealthier classes. Population of Mountain ecological region and Far western development region had high risk of infant mortality. Similarly, infant dying was higher for infants whose birth size, as reported by mothers, was very small and who has higher birth rank and short preceding birth interval. This study uniquely addresses both broader socioeconomic distal and proximate determinants side by side at the individual, household and community levels. For this, both comprehensive, long-term, equity-based public health interventions and immediate infant care programs are recommended. Infant mortality rate is defined as the risk of a live-born child to die before its first birthday. Infant mortality rates reflect economic and social conditions for the health of mothers and newborns, as well as the effectiveness of health systems . The cauMillennium Development Goal 4 aims for a two-thirds reduction in infant mortality by the year 2015 . Infant This study analyzed the secondary data from the nationally representative Nepal Demographic Health Survey (NDHS), 2011 accessed from the Measure Evaluation Demography Health Survey 2011 Nepal . The EnuThe Mosley and Chen conceptual framework for the study of child survival in developing countries Fig.  11] was was11] wThe outcome was infant death, which is the death of a live born infant in the first year of life. In this analysis, it was re-coded as a binary variable. The explanatory variables included community level distal socioeconomic, the household and individual level socioeconomic determinants and proximate determinants, covering maternal, infant, pre-natal, delivery, and post-natal factors in line with conceptual framework of study.People living in municipalities including towns and the capital city were considered as urban people and people living in villages or rural areas were considered as rural people. Development regions covered five administrative regions while ecological regions covered Mountain, Hill and Terai ecological zones.In this study, the main socioeconomic determinant is household wealth quintile (index). It is a method developed by the ORC Macro to measure the socioeconomic level for a household in a ranked order. It uses principal-component analysis based on respondents’ household assets, amenities, and services . In the +-birth rank. The preceding birth interval was grouped into two groups: less than 2-year and two or more years. These two variables are combined into one variable with categories [th or more birth order with 2-year or more of preceding spacing and 4+ birth rank with less than 2-years of preceding spacing.The proximate determinants include birth size, birth order and previous birth interval. Size at birth  was obtained by asking mothers. Birth rank was categorized into three groups: first, 2–3 birth rank and 4tegories . First rA synthetic cohort life table approach was used to calculate infant mortality rate. Data were weighted by sampling probabilities to represent the structure of Nepali population using weighting factors provided with the NDHS . Due to Frequency tabulations were used to describe the data, followed by the bivariate analysis using Chi-square tests and contingency table analyses to examine the association of all potential determinants on infant mortality without adjusting for other covariates. Prior to multivariate hierarchical logistic regression analysis, multi-collinearity between the variables was assessed and variables with multi-collinearity were not considered for the analysis. For example, parental education level and occupation were highly correlated with wealth index, so these variables were not considered in the analysis though they were significant. In addition, only those variables that were significant in the bivariate analysis were further analyzed using multivariate hierarchical logistic regression. A p-value less than 0.05 was considered as significant and odds ratios at 95 per cent confidence intervals were determined.2 was used to assess goodness-of-fit of logistic models. The Statistical Package for Social Science (SPSS 16.0 for Windows) software was used to analyze the data.Based on a conceptual framework describing the hierarchical relationships between different groups of variables, multivariate hierarchical logistic regression was used to assess the association of distal socioeconomic and proximate determinants on infant mortality after controlling other variables. In this approach the associations of more distal variables can be examined without improper adjustment by proximate or intermediate variables that may be mediators of the effects of more distal variables . At the The Nepal DHS 2011 was approved by the ethics review board of the ICF Macro International and the Ministry of Health and Population. All respondents were verbally informed before consenting to their participation. This research study ensured that their participation was voluntary and independent when answering and interacting with the interviewers. The researcher also maintained confidentiality of the information, which was received after permission from Measure DHS.This research included 5391 live-births, occurring within the 5 years preceding the survey. The characteristics of explanatory variables are given in Table Infant mortality rate was found to be 46 per 1000 live births between 2006 and 2010. IMR was 44 deaths per 1000 live births in the Terai ecological region, compared to 65 deaths per 1000 live births in the Mountain region. It was highest in the Far-western development region (66 deaths per 1000 live births) and lowest in the Western development region (40 deaths per 1000 live births). Similarly, IMR was higher (47 deaths per 1000 live births) in rural areas than in urban areas (40 deaths per 1000 live births). The IMR was 55 in the middle class households compared to 29 in the wealthiest households. IMR was 138, as compared with 38 in larger than average birth size. However, rest of the categories had no significant differences in infant mortality with average size at birth. IMR was 72 in 2–3 birth rank & = <2 years of birth interval and 81 in > =4 birth rank & = <2 years of birth interval category.Table Compared to infants born to mothers who have no formal education or are illiterate, the unadjusted odds of dying was higher for infants whose mothers have secondary and higher levels of formal education . Similarly, compared to infants born to fathers who have no formal education or are illiterate, the unadjusted odds of dying was higher for these infants compared to those whose fathers have secondary and higher levels of formal education . However, parental education level variables were not entered into the model simultaneously as they were found to be highly correlated to the wealth index  of experiencing infant mortality compared with the Terai plain/low land region. Similarly, the Far-western development region had a higher adjusted odds ratio  of experiencing infant mortality than with reference to the Western development region.The second model presents the results after adding the wealth index as a socioeconomic determinant of infant mortality. Even after inclusion of this variable in model 2, the association of community level determinants with infant mortality was retained for example, the adjusted odds ratio of infant death was 1.33 in mountain ecological region with reference to the Terai ecological region. Similarly, the adjusted odds ratio of infant death was 1.58 in the Far western development region with reference to the Western development region. Furthermore, in reference to infants of the Richest class, the adjusted odds ratio of infant dying was 1.66 (95 % CI: 1.00–2.74) in Middle class and 1.87 (95 % CI: 1.14–3.08) in Poorer class respectively.The third model presents the results after adding all the proximate determinants. The reduction of the significance level of socioeconomic determinants  after inclusion of the proximate determinants  indicates that distal determinants are important predictors for infant mortality. For example, in reference to infants of the Richest class, the adjusted odds ratio of infant dying increased to 1.72 (95 % CI: 1.03–2.87) in Middle class and 1.95 (95 % CI: 1.18–3.24) in Poorer class, respectively. Similarly, the association of proximate determinants with infant mortality was statistically significant. In reference to average sized babies, adjusted odds ratio of infant dying was higher for infants whose birth size according to the mother was very small . Similarly, the adjusted odds ratio of infant mortality for fourth or higher birth rank infants with a short preceding birth interval  was higher  compared to the second or third rank infants with longer birth intervals. A short birth interval of the second or the third rank infants also showed increased odds of infant deaths   has targeted to reduce infant mortality at 34 per 1000 live births .Similarly, differences in terms of regional variation were not statistically significant. Likewise, the findings of this analysis showed that sex of the infants did not influence the odds of dying but the literature shows females have lower odds of mortality than males during the first month of life –29. TherAll above-mentioned socioeconomic determinants operated through a common set of significant proximate determinants of infant deaths. These determinants were size of babies at birth and birth rank and birth interval. Smaller infant size at birth was found to be one of the strongest determinants of infant mortality. This finding is supported by other literature as well. Low birth weight was a strong predictor of neonatal mortality . Food-avFacility-based, population outreach, or home/family/community based antenatal, natal and postnatal care interventions have been proven to be effective to prevent infant deaths –36. TherThe analysis also found that there was no significant difference between age of the mother and infant mortality though there is a high prevalence of early marriage and early pregnancy in Nepal. In line with this finding, an analysis of the World Fertility Survey data  showed tThe identification of key determinants of infant deaths is important to provide guidance for the development of evidence-based focused interventions. In line with this need, National Health Policy, 2014 and Nepal Health Sector Program (2015–2020) have provisioned equity as a guiding principle of health programs. For this, as Buyana suggests, local government budgeting should be in a two-fold framework that combines both disease-based health needs and socio-economic needs . Thus, iThis paper included live-births, occurred within the 5 years preceding the survey. The associations of infant mortality with factors drawn from statistical analyses might lack a temporal relationship. This is due to the cross-sectional design used in Nepal Demographic Health Survey, 2011, thus limiting causal inference. For example, current poverty is a proxy for past poverty. Finally, the data for the Nepal Demographic Health Survey, 2011 was collected at the individual and household levels. For the present analysis, only crude community level indicators  were used.The analysis of NDHS data (2006 to 2010) in this paper demonstrated that socioeconomic determinants are associated with infant mortality in Nepal. Specifically, poorer and middle class people and people who reside in the Mountain ecological region and Far Western development region had high infant mortality. However, determinants like gender and urban/rural residence were found to be statistically insignificant.These socioeconomic determinants operated through a common set of proximate determinants such as size of babies at birth, birth interval or spacing associated with high infant deaths. Therefore, infant mortality is typically multi-factorial in causality and the cumulative consequences of interactions of social, economic and biological determinants, among others. Hence, findings point to address both socioeconomic and proximate determinants side by side. For this, comprehensive, long-term, equity-based public health interventions and immediate infant care programs are recommended. Moreover, this study recommends an advanced analytical study to explore the independent roles of key determinants of infant mortality in Nepal.


Autoimmune diseases frequently affect women of childbearing age. Gravidas with autoimmune diseases have a tendency to develop complications, such as pregnancy-induced hypertension in pregnancies complicated by systemic lupus erythematosus (SLE). The activity of autoimmune diseases is often exacerbated during pregnancy. The transplacental passage of autoantibodies may have an effect on the fetus or neonate; congenital heart block, in particular, is often irreversible. Therefore, increased vigilance is required for pregnancies complicated by autoimmune disorders.Herein the characteristics of newborns from mothers with autoimmune disorders were evaluated.Clinical manifestations were assessed in 40 newborn infants from 38 mothers with autoimmune disororders  who attended the Tokyo Women’s Medical University Hospital between January 2011 and November 2013. The following information deriverd from the patients’ charts was reviewed: maternal autoimmune disease, maternal autoantibodies, medications prescribed during pregnancy, gestational age (GA), and autoantibodies transferred to the infant.The average age of the mother at the time of the delivery was 33.7 years. Twenty mothers had anti-Ro/SS-A, 5 had anti-La/SS-B, and 4 had anti-RNP antibodies. Twenty-two gravidas (57.9%) were treated with oral glucocorticoids (GC) during pregnancy. The average dose of predonisolone and methylprednisolone was 10.8 mg/day (2-30 mg/day) and 8.0 mg/day (4-16 mg/day), respectively. Four gravidas with SLE used immunosuppresants . Seven of 38 gravidas (18.7%) were delivered by emergency cesarean section. The average GA at birth was 36 3/7 weeks (23 6/7-41 1/7 weeks) and 17 of 40 (42.5%) had preterm births. Four of 23 term newborns met the criteria for small for gestational age (SGA). The average birth weight was 2314±649 g , and included 23 low birth weight (LBW) newborns (57.5%) and 3 extremely LBW newborns (7.9%). When the gravida used GC during pregnancy, there were significantly higher rates of preterm births and SGA neonates (p<0.05). The birth weights and GAs were not significantly different between gravidas with and without SLE. The average GA and birth weight of newborns whose mothers were treated with GC was 35 5/7 weeks and 2131±704 g, respectively, suggesting that GC may be a factor that affects GA and birth weight. Autoantibodies were detected in 18 infants . One baby had congenital complete heart block. Rash, thrombocytopenia, and hyper-transaminasemia (ALT > 35 U/l) developed after hospital discharge in 1, 3, and 4 infants, respectively. With respect to neonates born to antiSS-A antibody-positive mothers, maternal GC use or non-use had no effect on the incidence of autoantibodies and infant symptoms. There were no infants who developed congenital heart block born to mothers who had used prednisolone during pregnancy.The probabilities of preterm delivery and LBW infants were higher and more remarkable in gravidas treated with GC. Further research is needed to better understand how maternal autoimmune disease activity influences the fetus and neonate. Continued observation by pediatricians is important until autoantibodies are cleared in the infant.None declared.


Low weight at birth is associated with obesity in later life. One hypothesis to explain such an association is that genetic variants that increase the risk of obesity also reduce fetal weight. Recently, obesity in adults was found to be associated with common variants of the fat mass and obesity-associated (FTO) gene. We examined the association between FTO polymorphisms and birth weight in a singleton, full-term birth cohort of 494 newborn-mother pairs without any complications.The risk alleles for obesity  were associated with low weight at birth. The mean differences per risk allele were −79 g  for rs9939609 and −84 g  for rs9930506. The level of association remained statistically significant after adjustment for the maternal risk allele and for variables usually associated with birth weight . In the follow-up, the allelic difference in weight was attenuated over time.The FTO variants that confer a predisposition to obesity later in life appear to be associated with low weight at birth. This finding favors the hypothesis of a common genetic denominator that predisposes to a low weight at birth and obesity in adults. Althoregnancy ,7. An alregnancy  is that Several independent, genome-wide, association studies have recently identified a strong correlation between fat mass and obesity-associated (FTO) polymorphisms and obesity-related parameters  in adults and children -11. In cIn the present study, we investigated the effect of two common FTO polymorphisms on birth weight in a singleton, full-term birth cohort without any maternal or newborn complications.The mothers and their newborns were consecutively recruited in the maternity ward during 2008 and 2009. The nurses of the maternal unit collected blood samples (cord blood) and recorded maternal (at entry) and newborn (at delivery) anthropometric parameters. Birth weight was measured to the nearest 1 g using a digital baby scale Seca model 727 , which includes a special damping system that allows for precise weighing, even if the newborn is restless. Two trained study nurses extracted data  from the obstetric medical records and interviewed the mothers to obtain data on smoking status and the genealogical tree (four generations).The selection criteria of the mother–newborn pairs were a Caucasian origin, eutocic delivery with cephalic presentation between the beginning of the 37th week and the end of the 41st week, singleton live birth, no maternal use of alcohol or illicit substances, no gestational complications, no diabetes, no congenital malformations or perinatal problems, and an Apgar score of ≥ 7 during the 1st minute and ≥ 9 by the 5th minute. Twenty-nine of the original mother–newborn pairs who were recruited by the nurses were excluded because of non-compliance with one or more inclusion criteria, difficulty of genotyping, or important missing data.For infants participating in the follow-up program of our hospital, body weight at 3, 6, 9, and 12 months was collected from the pediatric medical records. For the other children, we attempted to obtain data 1 year later by phone for the most recent weight of the child. The study protocol was approved by the Ethical Committee of the Hospital of Jolimont (ECO22) and informed written consent was obtained from all mothers before participation.We examined two single nucleotide polymorphisms (SNPs) of FTO  that were previously shown to be strongly associated with obesity-related parameters in large European populations . We wereSample size was estimated for testing the primary hypothesis that birth weight is different between newborns with different genotypes (BMI was not used because measurement of length is less reliable in newborns). Based on the weights obtained in the first 100 newborns (mean ± standard deviation: 3284 ± 347 g), the minimum size for each group that was required to detect a 5% difference (a priori considered as relevant) was estimated as 70, for a two-sided significance level of 0.05 and a statistical power of 80% . Based oThe analyses focused on birth weight. We examined associations separately for newborn and maternal genotypes, and then mutually adjusted one for the other, as well as for other non-genetic factors using multiple linear regression analysis . Adjustment for maternal genotype is important because of the strong association between maternal and newborn genotypes and our aim to establish whether any associations are primarily driven by newborn genetic variants independent of maternal genotypes. Other non-genetic factors that are known to affect birth weight were examined, including racial/ethnic origin , sex, gestational age, parity, cigarette smoking, maternal BMI, and weight gain during pregnancy. In all analyses, a per-allele additive genetic model was examined.The characteristics of the newborns and their mothers are shown in Table For both SNPs, newborn homozygotes for the risk allele weighed significantly less than newborn heterozygotes, and weighed even less than the homozygotes for the non-risk allele, with a clear gene-dose effect Table . The newBirth weight was correlated with sex , gestational age , parity , cigarette smoking , and maternal BMI , but not with Italian origin of the grandparents . There was only a trend toward statistical significance with weight gain during pregnancy . None of these factors was significantly associated with the risk alleles. In particular, maternal BMI was not significantly associated with newborn risk alleles, but it was significantly associated with maternal risk alleles . Additionally, maternal risk alleles were associated with newborn risk alleles. Adjustment for factors that were significantly (p < 0.05) associated with birth weight  only slightly decreased the strength of the associations between newborn risk alleles and birth weight, without affecting the statistical significance , has been found by Morgan et al. . Other sde novo lipogenesis [Although replication in independent samples is essential, our finding is compatible with the hypothesis  that coogenesis . High-riogenesis . The effogenesis , similarogenesis . In animogenesis . In humaogenesis , a materogenesis . Fetal Fogenesis ,46. HoweWe recognize that our study has some limitations. First, the associations observed in our study could be false positives. A false positive association is frequently caused by the confounding effect of population stratification when ethnicity or geographic origin is associated with the phenotype and genotype. We attempted to control for this type of bias by verifying and adjusting for the origin . Our sample size is small compared with the majority of genetic association studies. This resulted in a lower power to detect any associations with a high level of statistical significance. Calculation of statistical power using a mean difference per allele of 50 g showed that we had a 44% power to detect an association. Finally, birth weight is a simple measure that does not discern between fat mass and other components. Future studies need to investigate a more precise measure of fat mass in newborns using total body electric conductivity, dual energy x-ray absorptiometry, or air displacement plethysmography .In conclusion, the present study investigated 494 newborns with well-documented confounding factors that affect birth weight. After exclusion of pathological situations affecting birth weight, the FTO risk allele for obesity showed a significant, inverse association with birth weight. This association remained significant after correction for confounding factors and maternal FTO variants. This observation is compatible with the notion that genetic variants leading to obesity in later life may cause lower weight in fetal life, and supports a role for FTO in early growth.


In this study we assessed the mediation role of the gestational age on the effect of the infant’s congenital heart defects (CHD) on birth-weight.We used secondary data from the Baltimore-Washington Infant Study (1981–1989). Mediation analysis was employed to investigate whether gestational age acted as a mediator of the association between CHD and reduced birth-weight. We estimated the mediated effect, the mediation proportion, and their corresponding 95% confidence intervals (CI) using several methods.There were 3362 CHD cases and 3564 controls in the dataset with mean birth-weight of 3071 (SD = 729) and 3353 (SD = 603) grams, respectively; the mean gestational age was 38.9 (SD = 2.7) and 39.6 (SD = 2.2) weeks, respectively. After adjusting for covariates, the estimated mediated effect by gestational age was 113.5 grams  and the mediation proportion was 40.7% , using the bootstrap approach.Gestational age may account for about 41% of the overall effect of heart defects on reduced infant birth-weight. Improved prenatal care and other public health efforts that promote full term delivery, particularly targeting high-risk families and mothers known to be carrying a fetus with CHD, may therefore be expected to improve the birth-weight of these infants and their long term health. Congenital heart defects (CHD), also known as congenital cardiovascular malformations, are the most common type of structural birth defects and have substantial effects on pediatric mortality and morbidity , 2. The CHDs originate in the early weeks of pregnancy when the heart is forming and are usually diagnosed either prenatally or soon after birth, although some defects are not diagnosed until later in childhood or even in adulthood. Among the most consistently observed findings in newborns with CHD are reduced birth weight , 7–11 anThe aim of our study was to investigate the relationship between the CHD status of infants and their birth weight by considering gestational age as a mediating variable (mediator), i.e., a variable that “mediates” the relation between the predictor (an infant’s CHD status) and the outcome (birth weight). Using statistical mediation analysis, we considered gestational age as the mediator, we assessed its indirect effect on the relationship of interest, and we estimated the proportion of the effect of CHD status on infant birth weight accounted for by this mediator, taking into account infant and maternal covariates known to affect birth weight and CHD status.The Baltimore-Washington Infant Study was a population-based case–control study of the potential risk factors of CHD, conducted from 1981 to 1989 in Maryland, the District of Columbia, and Northern Virginia. The study enrolled all live-born infants with CHD and a randomly selected sample of controls that was representative of all regional live births. Controls were randomly sampled at birth hospitals in the region during the same time period. Parents of cases and controls were interviewed at home by trained personnel within one year of the infant’s birth and asked questions concerning medical and demographic/socioeconomic factors, as well as questions regarding cigarette smoking and alcohol consumption. Mothers reported the infant’s birth weight in pounds and ounces, which we converted to grams for this analysis. Gestational age was calculated from the mother’s report of the infant’s date of birth in relation to her reported expected delivery date . The diaFrom the extensive Baltimore-Washington Infant Study dataset, we selected variables that are known from prior reports to influence both infants’ CHD status and birth weight, as potential confounders for our study. These included an infant’s sex –26 and rIn mediation analysis, the main interest is to assess how much the effect of an exposure on a response variable is mediated through an intermediate variable, oftentimes called the mediator . Thus, tAfter obtaining parameter estimates from these three regression models, we estimated the mediated effect as the product of the estimate of the effect of the independent variable on the mediator from Model 2 with the estimate of the effect of the mediator on the dependent variable from Model 3 , 40–43. From an epidemiological point of view, in order to explore the mechanism of exposure-disease relations, it is important to assess the mediated (indirect) effect. A relevant way of expressing mediated effects is by considering the “proportion mediated”, which is the proportion of the total effect explained by a particular mediator –40, 48. The parents of all participants signed an informed consent agreement to participate in the original Baltimore-Washington Infant Study protocol, which was approved by the Instututional Review Board of the University of Maryland, Baltimore. The present analysis was performed on data abstracted without identifiers.p < 0.0001), confirming the expected deficit among the infants with CHD. The average gestational ages were found to be 38.9 (standard deviation = 2.7) and 39.6 (standard deviation = 2.2) weeks, respectively, and the difference was statistically significant (p < 0.0001). Infant sex, maternal education, and maternal cigarette smoking were distributed similarly in cases and controls (p > 0.05), whereas the case–control differences in infant race, maternal age, previous pregnancies, pre-pregnancy diabetes, and method of delivery were each statistically significant. Based on these results we explored these variables as potential covariates in the subsequent statistical models.The Baltimore-Washington Infant Study enrolled a total of 3,377 cases and 3,572 controls from 1981 to 1989. Fifteen cases and 8 controls were dropped from the analyses due to missing values. Characteristics of the study population are shown in Table p-values from the three regression models  are shown in Table p > 0.10; where p = 0.10 was used as statistical cut-point for the inclusion of interaction terms into the final models). Race of the infant, maternal age, education and parity were not statistically significant in any of the models considered and were not included in the final models. Two covariates – sex of the infant and maternal smoking status – were associated with infant birth weight, with or without gestational age in the model. In Model 1, the CHD status had a statistically significant detrimental effect on birth weight  and Fieller’s methods , and 40.7%  using the bootstrap method (Table p = 0.3342).We also investigated the mediation proportion for the two sexes separately. The mediation proportion was 45.2%  for males and 36.7%  for females; the difference in these two proportions was not statistically significant (p = 0.002) and had a lesser affect on controls , the use of caesarian section did not significantly affect the mean birth weights of infants in this study.In studying the association between an independent and dependent variable, there are situations in which the independent variable affects the outcome variable both directly and indirectly, by influencing a third variable (the mediator) which in turn influences the dependent variable. Thus, the relationships between variables are often complex and cannot be well modeled by a single regression equation, while consideration of a mediator allows researchers to more fully explore the separate and combined effects of interest. The problem of CHD as a potential cause of reduced birth weight in affected infants is amenable to this approach because birth weight itself is highly dependent on gestational age, which in turn may be influenced by CHD to some extent. Consistent with this view, several studies have previously documented the association of CHD in newborns with their decreased birth weight and gestational age, relative to unaffected infants, but to our knowledge, the mediation effect has not been directly assessed in prior published research , 13. AddThis study has several strengths and limitations. The data is from the Baltimore-Washington Infant Study, which had achieved complete ascertainment of cases in the region and a population-based random sample of controls; therefore selection bias is extremely unlikely as an explanation for our findings. On the other hand, we cannot rule out the possibility of recall bias of infant birth weight, nor random (non-systematic) errors in recalling the weight or the estimated date of delivery. There are also statistical issues that should be considered in evaluating our results. We confined our analysis to the linear effects of variables on birth weight, but future work should consider possible non-linear effects as well. While our use of regression models permitted us to assess these mediation effects, the technique can become unwieldy for more complex situations , 44, 54.In summary, we used data from a population-based epidemiological study to evaluate the mediation effect of gestational age in explaining the relationship between CHD and decreased birth weight of infants. The results suggest that this mediation accounts for a substantial proportion (about 41%) of the latter association, which was not apparent in previous research in this field due to the limitations of conventional statistical association methods. Further research is needed to confirm these findings in other populations and to explore even more complex methods such as structural equation modeling , 56.