Although a number of studies have investigated correlations of maternal age with birth outcomes, an extensive assessment using age as a continuous variable is lacking. In the current study, we estimated age-specific risks of adverse birth outcomes in childbearing women.National population-based data containing maternal and neonatal information were derived from the Health Promotion Administration, Taiwan. A composite adverse birth outcome was defined as at least anyone of stillbirth, preterm birth, low birth weight, macrosomia, neonatal death, congenital anomaly, and small for gestational age (SGA). Singletons were further analyzed for outcomes of live birth in relation to each year of maternal age. A log-binomial model was used to adjust for possible confounders of maternal and neonatal factors.In total, 2,123,751 births between 2001 and 2010 were utilized in the analysis. The risk of a composite adverse birth outcome was significantly higher at extreme maternal ages. In specific, risks of stillbirth, neonatal death, preterm birth, congenital anomaly, and low birth weight were higher at the extremes of maternal age. Furthermore, risk of macrosomia rose proportionally with an increasing maternal age. In contrast, risk of SGA declined proportionally with an increasing maternal age. The log-binomial model showed greater risks at the maternal ages of <26 and > 30 years for a composite adverse birth outcome.Infants born to teenagers and women at advanced age possess greater risks for stillbirth, preterm birth, neonatal death, congenital anomaly, and low birth weight. Pregnancies at advanced age carry an additional risk for macrosomia, while teenage pregnancies carry an additional risk for SGA. The data suggest that the optimal maternal ages to minimize adverse birth outcomes are 26∼30 years. There is a trend in increasing maternal age for childbearing worldwide. Changes in maternal age may have impacts on birth outcomes Some studies that examined the relationship between maternal age and adverse birth outcomes used data from regional samples or from a limited number of medical institutions The study protocol was approved by the Research Ethics Committee of the National Health Research Institutes in Taiwan. All records of participants were anonymized and de-identified prior to analysis. Targets of this retrospective population-based study were all births from 1 January 2001 to 31 December 2010 in Taiwan.Maternal and neonatal data were derived from the Birth Notification System (BNS), a database established by the Health Promotion Administration, Ministry of Health and Welfare, Taiwan . MedicalAdverse birth outcomes were measured using the following 7 outcomes: stillbirth, preterm birth, neonatal death, congenital anomaly, low birth weight, macrosomia, and SGA . A compoMaternal age was defined as the age at delivery, which was calculated by subtracting the maternal birthday from the neonatal birthday. Women with a delivery age of <20 years old were classified as teenage mothers.All births were included when estimating the stillbirth and composite adverse birth outcome. Otherwise, cases of stillbirths and multiple births were excluded when measuring the correlations of maternal age with preterm birth, low birth weight, neonatal death, macrosomia, congenital anomaly, SGA, and delivery mode.Statistical analyses were conducted using a commercially available program . Categorical variables were analyzed using a chi-squared test. Relative risk with 95% confidence intervals (CI) was expressed. For comparison between groups with quantitative variables, the null hypothesis that there was no difference between each group was tested by a one-way analysis of variance (ANOVA). Population attributable fraction (PAF) was estimated by the following formula:PAF  =  [P(RR-1)]/[P(RR-1)+1]P: proportion of composite adverse birth outcomes; RR: risk ratiop<0.05. Absolute risk difference and 95% CI were adjusted for the control variables.A log-binomial model  was used to estimate the risk of adverse birth outcomes in relation to maternal age after adjusting for possible confounders of maternal and neonatal factors – including pregnancy-related disorders, birth region, parity, obstetric complications, ethnicity, birth year, sex at birth, congenital anomaly, neonatal death, Apgar score, delivery mode, gestational age, and birth weight. All covariates are defined in p<0.001). Seven adverse birth outcomes are illustrated year by year in the Data on 2,123,781 births were collected in Taiwan from 2001 to 2010. We excluded 30 births with incomplete data, leaving 2,123,751 births for the analysis. There were 20,489 stillbirths (0.96%). Among 2,045,748 singleton live births, 62,839 neonates (3.07%) were born to teenage mothers and 243,624 neonates (11.91%) were born to mothers of ≥35 years of age. Socio-demographic data of the study population are summarized in The risk of a composite adverse birth outcome was highest at ages of ≤14 years, then declined to an age of 27 years, and then steadily increased to ages of ≥44 years . In addiThe highest risk of stillbirths was for ages of ≤14 years, then declined to ages of 22∼29 years, and then steadily increased to ages of ≥44 years. In particular, the risk of stillbirths at ages of ≥44 years was significantly lower than that at ages ≤15 years (p<0.001). The univariate analysis showed that pregnant women aged younger than 22 years or older than 29 years carried a greater risk for stillbirths compared to those aged 27 years.Pregnant women aged <25 years or > 28 years carried a greater risk for preterm birth compared to those aged 27 years. The risk of preterm births was highest at ages of ≤14 years, then declined to ages of 25∼28 years, and then steadily increased to ages of ≥44 years.Pregnant women aged <27 years or > 32 years carried a greater risk for having a low birth weight infant compared to those aged 27 years. The risk of low birth weight neonates was highest at ages of ≤14 years, which declined to ages of 27∼32 years, and then increased to an age of 43 years and older.Risk of neonatal death was significantly higher at the extremes of maternal age. The risk of neonatal death was highest at ages of ≤14 years, which declined to an age of 28 years, and then increased to ages of ≥44 years.There was a significant correlation of macrosomia with maternal age (p<0.001): risk of macrosomia was significantly lower at younger age and higher at older age. The lowest and highest risks of macrosomia were at ages of ≤14 and 43 years, respectively.Congenital anomaly was more common in stillbirths than live births (data not shown). Risk of congenital anomaly was significantly higher at extreme maternal ages.There was a significant correlation of SGA with maternal age (p<0.001): risk of SGA was higher at younger age and lower at older age. The highest and lowest risks of SGA were at ages of ≤14 and 36 years, respectively.p<0.001): risk of CS was significantly lower at younger age and higher at older age (data not shown).In total, 686,491 births were by Cesarean section (CS) (33.56%). There was a significant correlation of delivery mode with maternal age .The current evaluation illustrates correlations of maternal age with the following 7 adverse birth outcomes – stillbirth, preterm birth, neonatal death, congenital anomaly, macrosomia, SGA, and low birth weight – for over 2 million deliveries. Although previous investigations established maternal age to be an important factor in relation to adverse birth outcomes, data were less extensive in terms of estimates of the risk. In most previous studies, maternal age was categorized or dichotomized for the analysis. Our study differs from those studies in that we classified each year of age as an age-specific variable to allow for a more-precise and -realistic analysis of associations between maternal age and adverse birth outcomes. In addition, we provide a wealth of information regarding the public health magnitude of the issue by showing the risk of composite outcomes. Furthermore, we adjusted for control variables by log-binomial model and excluded some population (ex. multiple births) to minimize possible bias. Findings of our study demonstrate that women bearing children early or late in life possess a greater likelihood for composite adverse birth outcomes – including stillbirth, preterm birth, neonatal death, congenital anomaly, and low birth weight. In addition, pregnant women at advanced age are more likely to have CS and macrosomic infants. Furthermore, teenage pregnancies carry additional risk for SGA. To our knowledge, this is the first survey to comprehensively analyze age-specific risks of adverse birth outcomes.Our study identified specific maternal ages which are at lower risk for a number of adverse birth outcomes during 2001∼2010 in Taiwan. We verified optimal ages for having a live birth as 22∼28 years. These data are consistent with previous reports showing greater risk of stillbirth at the extremes of reproductive age In this study period, optimal ages of childbearing women to minimize preterm birth were 23∼28 years. In addition, optimal maternal ages to avoid low birth weight were 27∼33 years. These findings are largely consistent with those of other relevant studies showing that both teenagers and women of advanced maternal age are at greater risk for having preterm birth It is well acknowledged that teenage pregnancies are at increased risk for adverse birth outcomes The mechanisms underlying the increased risk for adverse outcomes with extremes of maternal age are uncertain. Effects may result from direct biological changes or environmental impacts Our study illustrated that the risk of CS was proportional to maternal age. This was probably because of an increasing propensity to request cesarean delivery with maternal age Our study showed a proportional rise of macrosomia with an increase in maternal age. This finding is consistent with previous reports There are limitations to this study. First, our samples were Asian and therefore cannot represent other races Our analyses illustrate a clear trend in reduced and delayed childbearing, which might have been driven by educational and labor gains of women over the last few decades. Women of childbearing age are confronted with the dilemma of choosing education, career, or pregnancy. Policy makers should consider designing certain feasible interventions to reduce the delivery age in an attempt to minimize adverse birth outcomes.The current study has measured adverse birth outcomes in relation to maternal reproductive age ranging from teenage to advanced age in a nationwide population-based setting. Our results demonstrate extremes of maternal age carry greater risks of stillbirth, preterm birth, neonatal death, congenital anomaly, and low birth weight with maternal age. In particular, pregnancies at very young age (<16 years) are at the greatest risk. The most optimal maternal ages to prevent overall adverse birth outcomes are 26∼30 years. Specifically, optimal maternal ages to minimize stillbirth, preterm birth, and low birth weight are 22∼28, 23∼28 and 27∼33 years, respectively. Our data highlight an imperative need to devise interventions to reduce adverse birth outcomes for pregnancies of teenagers and women of advanced age.S1 FigureRates of adverse birth outcomes among 2,123,751 births.(TIF)Click here for additional data file.


There was no effect of MM on concentrations of maternal hPL or PGH, or cord insulin, IGF-1, or IGFBP-1. However, among pregnancies of female offspring, hPL concentration was higher by 1.1 mg/L in the third trimester ; and among women with height <145 cm, insulin was higher by 59%  in the MM vs. IFA group. Maternal hPL and cord blood insulin and IGF-1 were positively, and IGFBP-1 was negatively, associated with birth weight z score and other measures of birth size . IGF-1 was inversely associated with gestational age (p<0.05), but other growth factors were not associated with gestational age or preterm birth. Prenatal MM supplementation had no overall impact on intrauterine growth factors. MM supplementation altered some growth factors differentially by maternal early pregnancy nutritional status and sex of the offspring, but this should be examined in other studies.Prenatal multiple micronutrient (MM) supplementation improves birth weight through increased fetal growth and gestational age, but whether maternal or fetal growth factors are involved is unclear. Our objective was to examine the effect of prenatal MM supplementation on intrauterine growth factors and the associations between growth factors and birth outcomes in a rural setting in Bangladesh. In a double-blind, cluster-randomized, controlled trial of MM vs. iron and folic acid (IFA) supplementation, we measured placental growth hormone (PGH) at 10 weeks and PGH and human placental lactogen (hPL) at 32 weeks gestation in maternal plasma (n = 396) and insulin, insulin-like growth factor-1 (IGF-1), and IGF binding protein-1 (IGFBP-1) in cord plasma (n = 325). Birth size and gestational age were also assessed. Early pregnancy mean (SD) BMI was 19.5 (2.4) kg/mNCT00860470ClinicalTrials.gov  Intrauterine growth restriction remains a widespread public health issue across the globe . Pooled Little is known about what mechanisms underlie such an effect, but hormones and growth factors that regulate fetal growth are clear targets for research , 5. InsuRelationships between micronutrients and growth factors have been examined in animal models and cellular research. Zinc and calcium have well-established roles in insulin synthesis and secretion  and circThere is limited understanding of the potential for prenatal micronutrient supplementation to enhance in utero growth through changes in growth factors, especially in settings with high rates of malnutrition. Thus, we examined the impact of MM on key fetal growth factors in the context of a randomized trial of MM compared to IFA in rural Bangladesh. Further, we assessed the relationship between these growth factors and placental weight, infant size, and gestational age at birth.12 (2.5 mg), vitamin B6 (1.9 mg), vitamin C (85 mg), zinc (12 mg), iodine (220 μg), copper (1000 μg), and selenium (60 μg). The trial was conducted in rural northwestern Bangladesh where the ~435 km2 study area was divided into 596 sectors of comparable size that were used as units of randomization  participating in a double-masked, cluster-randomized controlled trial  of daily MM or IFA (standard of care) supplementation to reduce infant mortality . The MM A substudy was conducted in 64 contiguous sectors (32 per group) to assess the changes in maternal vitamin and mineral status due to supplementation. The area was purposely selected to be accessible by road but similar to the parent trial across several socio-demographic and geographic factors. Maternal venous blood was obtained at 10 and 32 weeks of gestation, among numerous other measurements. Approximately half of this substudy area (31 sectors) was selected for an additional intensive home-based cord blood collection protocol. A primary outcome of this study was concentrations of maternal and fetal growth factors, assessed at 10 and 32 weeks gestation  and at birth . From February 2009 to April 2010, newly pregnant women in the selected substudy area were asked for consent  for cord blood collection after enrolling in the parent study at a median of 9.6  weeks gestation. Final births from enrolled women, and the end of follow up for this substudy, occurred in November 2010. Those >28 weeks gestation at the time of enrollment were excluded from the cord blood substudy. Sample size (n = 310 cord blood samples) was set to detect differences in micronutrient status.Local field workers met women weekly to provide the micronutrient supplements from enrollment to 3 months postpartum; supplements were self-administered daily. Trained data collectors measured women in their homes at the time of substudy enrollment. A questionnaire was used to collect self-reported data on pregnancy history, diet, strenuous work, morbidities, socio-demographics, and other maternal and household factors. Study staff trained to do ultrasound assessments  measured crown-rump length transabdominally at <15 wk to estimate gestational age; date of last menstrual period was used for gestational age estimation if ultrasound was unavailable (n = 44). Before supplementation (~10 wks) and at 32 weeks gestation, the technicians collected 7 mL of maternal venous blood. A drop was used to measure hemoglobin on the spot with a HemoCue photometer . Blood collection tubes were immediately stored in padded, insulated cold boxes with ice packs until technicians returned to the central lab for processing. Processing occurred approximately 1 to 6 hours after collection due to field workers traveling to several homes in difficult to reach rural areas in the same day before returning to the lab.Because over 90% of women deliver at home in this rural setting, we developed an extensive labor tracking system to allow women or a close family member or friend to notify us of the beginning of labor. Trained nurse midwives were on call 24 hours a day to go to the home (or other delivery site) when notified. Immediately following birth and separation of the infant and placenta, cord venous blood (7 mL) was collected by the trained nurse midwives from the placental side with a syringe; transferred to a 7 mL blood collection tube; put in a padded, insulated cold box with ice packs; and transported to the field-based laboratory for processing. Cord blood processing typically occurred 1–2 hours after collection. Maternal and cord blood tubes were centrifuged and separated plasma was measured into multiple aliquots and immediately stored in liquid nitrogen until received at our US laboratory. Thereafter, plasma was stored at -80°C until assayed, without freeze-thaw cycles.At birth, we weighed trimmed, drained placentas  and newborns  on separate scales . Data wan for final analysis).We have previously published analyses using this same substudy cohort of 500 enrolled women, and some additional details of the methods, not relevant here, can be found in those papers , 22. ThehPL was run using ELISA kits purchased from IBL-America , with an inter-assay CV of 17%. Concentrations were too low at 10 weeks to be reliably detected and thus only 32 week samples were assayed. PGH was run using ELISA kits purchased from MyBiosource.com, with an inter-assay CV of 14% (assayed in both 10 and 32 week gestation samples). IGF-1 was analyzed using an automated chemiluminescent immunoassay system . Inter-assay CV was 9.5% for pooled plasma run as a quality control. Insulin and IGFBP-1 were each assessed using a high sensitivity ELISA kit . Insulin had an inter-assay CV of 6.2% for a pooled plasma sample; IGFBP-1 had an inter-assay CV of 4.6% for a quality control sample provided by the manufacturer. Lab technicians were blinded to supplementation group.th percentile) and birth weight z scores were calculated by a gestational age- and sex-specific birth weight reference [Small-for-gestational-age ; extreme statistical outliers  or low undetectable concentrations for insulin (n = 2) or IGF (n = 18), which did not meaningfully change results and thus are not presented. As well, there were no meaningful differences in results when restricting analyses of maternal samples to pregnancies ending in singleton, live births (n = 387). All data analysis was conducted with Stata 13.0  and statistical significance was considered at p<0.05.The trial and substudies received ethical approvals from the Institutional Review Board (IRB) at the Johns Hopkins School of Public Health, Baltimore, MD and the Bangladesh Medical Research Council (BMRC), Dhaka, Bangladesh. The current study was approved on December 11, 2008 by the Johns Hopkins’ IRB and on December 1, 2008 by the BMRC. The parent trial is registered on clinicaltrials.gov (NCT00860470). The process for registration began in 2007; however, the official release of the trial registration occurred after the study began due to numerous issues related to accurately documenting this large, community-based field trial on a site designed for patient-based, US clinical trials. It required substantial time and troubleshooting between the study investigators, the Johns Hopkins’ compliance officer, and the clinicaltrials.gov curators. A remnant of this process still remains with our major outcomes of infant mortality, preterm birth, and low birth weight being flagged as unrecognized conditions.Maternal plasma PGH increased from a median (IQR) 2.6  ng/mL at 10 weeks gestation to 117.0  ng/mL at 32 weeks. Maternal hPL was 9.4  mg/L at 32 weeks. Cord plasma insulin was 2.8  μIU/mL, IGF-1 was 32.2  μg/L, and IGFBP-189.0  μg/L at birth. IGFBP-1 was negatively correlated with IGF-1 (r = -0.41) and insulin (r = -0.28) and IGF-1 was positively correlated with insulin . PGH at 32 weeks was negatively correlated insulin (r = -0.14) PGH and hPL were not correlated with each other and hPL was not correlated with any of the fetal growth factors examined in cord blood .There was no difference in mean concentrations of hPL at 32 weeks between those receiving MM vs. IFA when accounting for the cluster design of the trial . We obseExamination of possible interacting variables revealed that among short women (<145 cm), insulin was 59% higher  in cord blood of the newborns of women consuming multiple micronutrients compared to iron-folic acid . We also found a 12% increased risk of SGA and 16% increased risk of LBW for each standard deviation increase in IGFBP-1. There was no association between maternal hPL, maternal PGH, or cord insulin concentrations and risk of SGA or LBW, or any associations between the fetal growth factors assessed and risk of preterm birth  compared to IFA, the continued prenatal standard of care worldwide . EvidencNow research is needed to explain the modest but consistent improvement in birth weight and reduction in risk of SGA in response to the UNIMMAP or similar supplements in populations worldwide. Thus, we aimed to elucidate mechanisms of the MM supplementation impact on fetal growth, in the context of an undernourished setting, by examining the impact of micronutrient supplementation on hPL, PGH, insulin, IGF-1, and IGFBP-1, all which have been shown to impact birth size. We did not find an overall effect of MM on these growth factors and the power to detect such differences was low, at ≤50% for all outcomes. Aside from the sample size limitation, we postulate that the true lack of a difference between these factors by supplement group may be best explained by 1) the lack of an impact on fetal growth shown in the parent trial in which overall birth weight increments were attributed to the two days of longer time in utero ; and 2) To our knowledge, this is one of two studies that examined the link between maternal MM supplementation, cord blood growth factors, and infant outcomes. In Burkina Faso, where birth weight was 2.9 kg, cord blood insulin was 2.1 and 1.9 μIU/mL and IGF-1 was 29 and 31 ng/mL in the IFA and MM groups, respectively , showing no impact on UNIMMAP vs. IFA on cord hormones . Yet theThere are many potential and plausible modifiers of the MM effect on fetal growth factors that have been reported previously. Here we examined maternal early pregnancy (as a proxy for prepregnancy) BMI, height, age, and parity and infant sex. We found that maternal height modified the effect of MM on insulin such that only in the “short” women, presumably chronically malnourished, did the supplement have an impact over IFA. Our analysis by low BMI did not show such an interaction, suggesting that supplementation may be benefiting more the chronically, rather than acutely undernourished women. Of note, in these home births and even in the small number of clinic births, women did not have routine medical procedures during labor such IVs with glucose that could have impacted cord insulin concentrations.Further, we found an infant sex interaction such that hPL was higher in mothers receiving MM that were carrying female fetuses. This finding is difficult to interpret yet contributes to the wide range of findings concerning sex differences in response to maternal nutritional status during pregnancy –29. We sFinally, we confirmed that these growth factors had the expected positive  and negative (IGFBP-1) relationships with fetal and placental growth, even in a population with high rates of undernutrition. IGF-1 and IGFB-1 were related not just with birth size, but also fetal growth, as both factors were strongly associated with the outcome of SGA. Our unanticipated finding that IGF-1 was inversely related to gestational age is likely due to the fact that IGF-1 spikes in the third trimester and then declines until delivery .Our study had a small sample size that was powered to detected differences in micronutrient status rather than growth factors. Testing MM compared to IFA presented several limitations including the lack of a placebo group and not being able to separate specific micronutrient effects. As well, we measured only five fetal and placental growth factors in a complex system of fetal growth regulation. Our blood samples were not collected according to ideal lab protocols due to extremely challenging field conditions. Yet, they were collected with the utmost care, were treated in a manner designed to maintain the integrity of the analytes of interest, and represent a rare opportunity to study these factors in the context of home births in a high nutritional-risk population.Fetal growth occurs as a result of complex interactions throughout gestation between the maternal environment, including her nutrition and health, and endocrine factors of maternal, placental and fetal origin. In this study, growth factors were associated, as expected from other contexts, with size at birth and placental weight, demonstrating the utility of these biomarkers in settings with high rates of malnutrition. Daily, routine prenatal MM supplementation did not differentially impact, overall, fetal growth factors in maternal and cord blood compared to supplementation with IFA alone. This finding however was consistent with the findings of the parent trial–where MM affected length of gestation but not fetal growth. Differences in growth factors were observed in subgroups, including the nutritionally at-risk subgroup of stunted women, and these interactions should be explored in future human trials. Other potential pathways connecting micronutrient supplementation to fetal growth should also be explored.S1 CONSORT Checklist(PDF)Click here for additional data file.S1 ProtocolParent Trial Protocol and Amendment.(ZIP)Click here for additional data file.


Esophageal duplication cyst is a rare congenital anomaly resulting from a foregut budding error during the fourth to sixth week of embryonic development. Cervical esophageal duplication cysts are very rare and may cause respiratory distress in infancy. A full-term newborn girl who was born by normal delivery was transferred to our hospital because of swelling of the right anterior neck since birth. Cervical ultrasonography showed a 40 × 24 × 33 mm simple cyst on the right neck. Tracheal intubation was required at 2 weeks of age because of worsening external compression of the trachea. Fine-needle aspiration cytology revealed the existence of ciliated epithelium. At 1 month of age, exploration was performed through a transverse neck incision. The cyst had a layer of muscle connected to the lateral wall of the esophagus. Histopathological diagnosis was a cervical esophageal duplication cyst. We describe the clinical features of infantile cervical esophageal duplication cysts based on our experience of this rare disease in a neonate, along with a review of 19 cases previously reported in literature. Esophageal duplication cysts are rare congenital cystic masses resulting from an error in foregut budding in the developing embryo and with a reported incidence of 1 in 8200 autopsies .–29.13–29This report described a very rare case of a neonate with cervical esophageal duplication cyst that was found at birth. Airway stenosis was progressive and required airway control with tracheal intubation. Exploration at 1 month of age led to the resolution of airway stenosis.According to 19 case reports and our own, respiratory distresses caused by external compression and deviation of the trachea commonly occur in infantile cervical esophageal duplication. We assume that cervical esophageal duplication cysts should be included in the differential diagnosis of an infant with respiratory distress and airway stenosis. Definitive diagnosis is based on histopathological findings, and surgical removal of the cyst is the treatment of choice.Written informed consent was obtained from the patient for publication of this case report and any accompanying images. A copy of the written consent is available for review by the Editor-in-Chief of this journal.


Few studies in low-income settings analyse linear growth trajectories from foetal life to pre-adolescence. The aim of this study is to describe linear growth and stunting from birth to 10 years in rural Bangladesh and to analyse whether maternal and environmental determinants at conception are associated with linear growth throughout childhood and stunting at 10 years.adj 2.93, 95% CI: 2.06–4.20), mothers with no education  or those conceived in the pre-monsoon season .Pregnant women participating in the MINIMat trial were identified in early pregnancy and a birth cohort (n = 1054) was followed with 19 growth measurements from birth to 10 years. Analyses of baseline predictors and mean height-for-age Z-scores (HAZ) over time were modelled using GLMM. Logistic regression analysis was used to investigate the associations between baseline predictors and stunting (HAZ<-2) at 10 years. HAZ decreased to 2 years, followed by an increase up to 10 years, while the average height-for-age difference in cm (HAD) to the WHO reference median continued to increase up to 10 years. Prevalence of stunting was highest at 2 years (50%) decreasing to 29% at 10 years. Maternal height, maternal educational level and season of conception were all independent predictors of HAZ from birth to pre-adolescence (p<0.001) and stunting at 10 years. The highest probability to be stunted at 10 years was for children born by short mothers (<147.5 cm) (ORHeight growth trajectories and prevalence of stunting in pre-adolescence showed strong intergenerational associations, social differentials, and environmental influence from foetal life. Targeting women before and during pregnancy is needed for the prevention of impaired child growth. Child undernutrition remains one of the main global health challenges in low and middle-income countries (LMIC) . StuntinEarly linear growth retardation is associated with future negative health consequences, such as impaired cognitive development, reduced economic productivity in adulthood, unfavourable maternal reproductive outcomes and a risk of development of non-communicable diseases . UnderstAlthough the dynamics and timing of early growth faltering in low- and middle-income countries are well documented ,15 most In this paper we aim to describe linear growth trajectories, using both height for age scores and absolute height difference from reference median, and the occurrence of stunting from birth to 10 years of age in a rural Bangladeshi setting. We also analyse whether important stunting determinants present from early life, representing maternal malnutrition , social differentials , and early environment (season of conception) continue to be associated with linear growth trajectories throughout childhood and with stunting at ten years of age.Children of mothers participating in the MINIMat trial  were followed from birth to 10 years of age. MINIMat was a factorial randomized trial primarily evaluating the effect of an early invitation to prenatal food supplementation  combined with multiple micronutrient supplementation  to pregnant women and the effect on anaemia in pregnancy, birth weight and infant survival. The reported main effects of the interventions so far are improved infant and under five survival and reduced stunting at 4.5 years ,20. The Community health research workers with at least ten years of education interviewed the mothers in their homes at enrolment. They used structured questionnaires that included socioeconomic characteristics (SES), parental education and pregnancy history.Home visits were followed by clinical visits at local sub-centers run by the icddr,b in the Matlab area. Maternal weight and height were measured at around 8 weeks of gestation using an electronic scale , with a precision of 0.10 kg and a stadiometer to the nearest 0.1 cm. The children’s weight and length/height were measured at birth and after that every month up to one year and then every three months up to 24 months and then again at 4.5 and 10 years. Most birth anthropometry was measured within 72 h of birth. However, measurements were made even if the new-borns were reached after 72 hours. Measurements taken from 24 hours up to 30 days after birth were adjusted using an SD score transformation, assuming that infants remain in the same relative position in the anthropometric distribution during this period . The recThe socioeconomic characteristics of the households were determined using a continuous household asset score previously generated in this population including data on land ownership, the construction materials of house walls, ownership of household assets etc. [Maternal height reflects the mother’s nutritional status accumulated through nutritional, social and environmental exposures and is a convenient indicator for assessing intergenerational associations. Tertiles of mothers’ height were created from the initial sample of 4436 women . Height data were missing from two mothers, and these were excluded from analysis. Maternal education has been proven to be a strong determinant of stunting and is associated with caregiving behaviour, feeding practices, and age at first pregnancy, etc. [Season of conception reflects the environmental exposures of the foetus and the mother. The variable was calculated based on LMP date and categorized after Bangladesh’s three distinct seasons; the hot pre-monsoon season from March through May, the rainy monsoon season from June through October and the cool and dry winter from November through February [ets etc. . Tertilecy, etc. ,23,24. TFebruary .height for age z-scores (HAZ) and the difference in height in cm (HAD) from reference median were calculated. HAZ was calculated using the program WHOAnthro from birth to 4,5 years and AnthroPlus at ten years of age using WHO growth references for 0–5 and 5–19 years respectively [To characterize growth, both ectively ,27. UsinWritten informed consent was obtained from all parents of participating children . The Ethical Review Committee at the International Centre for Diarrhoeal Disease Research, Bangladesh and the Regional Ethical Review Board in Uppsala, Sweden approved the study .An initial dataset was selected with those having height measurements from birth, 24 months and 4.5 and 10 years. Data were linearly interpolated if one or maximum two consecutive measurements were missing in the age interval between birth and 2 years of age. For example, if the 4-month height measurement was missing, it was imputed by linear interpolation of the 3- and 5-month heights. If two consecutive values were missing, the values were imputed from the weighted average of the previous and subsequent measurements.2) in the models. As the time points were not equally spaced, the correlation structure of the repeated measurements was modelled using an autocorrelation structure of order 1, with a continuous time covariate (CAR)[Generalized linear mixed models (GLMM) were used to model repeated measurements on the continuous outcome HAZ over the 19 different time points from birth to pre-adolescence. In the models Time (stating the 19 time points) was treated as nested under the random factor ID (the unique identifier for each child), thus allowing for different intercepts and slopes on the HAZ scores over time for each child. We found that the trend of HAZ over time was best explained by including a second order polynomial (Timeate (CAR). In the ate (CAR). The lonate (CAR).There were 3625 live born children by the 4436 women enrolled into the MINIMat trial. Losses before measurement of birth anthropometry included early foetal loss (n = 347), stillbirth (n = 89), outmigration (n = 188), and refusal to participate (n = 129). Out of the one-year birth cohort born from April 2002 to June 2003 selected for the 10-year follow-up 108 had no birth anthropometry and were excluded from analysis. Losses from follow-up after birth anthropometry up to 10 years of age included outmigration (n = 208), refusal to participate (n = 6) and death (n = 96). Of the 1187 children with anthropometry from birth and 10 years 134 children were excluded because of no height measurement from either 24 or 54 months resulting in 1054 children available for longitudinal analysis . The ranHeight for age scores (HAZ) and the difference in height in cm from reference median (HAD) from birth up to 10 years are shown in The prevalence of stunting at birth was 16% and increasing to 54% at 2 years of age . By 4.5 adj (95% CIadj: -0.71 to -0.46) lower than children to the tallest mothers. Children born to the shortest mothers also had a higher prevalence of stunting at all ages Click here for additional data file.S2 Table(PDF)Click here for additional data file.S3 Table(PDF)Click here for additional data file.S1 DataHeight for age scores for children participating in the MINIMat trial from birth to ten years of age, the final sample. N = 1054. (CSV)Click here for additional data file.


Estimates of the period mean age at first birth are readily available for countries with accurate vital statistics . In contrast, in most developing countries vital statistics are lacking or incomplete and estimates of the period mean age at first birth are therefore often unavailable. The Demographic and Health Surveys (DHS) program provides a large set of demographic and health statistics for many developing countries, but not the mean age at childbearing or the mean age at first birth.We propose two different methods for the estimation of the period mean age at first birth from information collected in DHS surveys. The first method is the same as the one used in populations with accurate vital statistics and is based on a weighted average of single year of age first birth rates. The second is the singulate mean age at first birth.A comparison of the two estimates obtained from the latest surveys in 62 countries shows excellent agreement in countries in which there is no evidence of a rise in childlessness. But, as expected on theoretical grounds, there is less agreement in populations that have experienced an increase in the proportion childless.Based on these results, we recommend the first method. The measure is relatively straightforward to calculate and, since it refers to recent births, is presumably more accurately reported than indicators based on events that occurred in the more distant past. This measure makes it possible for the first time to assess recent trends in the onset of childbearing in developing countries with multiple DHS surveys and to compare recent period estimates of the mean age at first birth among countries.The online version of this article (doi:10.1186/s12963-015-0058-9) contains supplementary material, which is available to authorized users. Becoming a parent for the first time is one of life’s most important and influential events. It signals the onset of the responsibility for insuring the well-being and success of one’s offspring and of the next generation. For women, the age at which they have a first birth can have implications for schooling, labor force participation, and overall family size . Early cEstimates of both cohort and period mean ages at first birth are available for countries with reliable vital statistics. For example, EUROSTAT  and the For many analytic purposes estimates of period measures are of greatest interest because, in contrast to cohort medians, they allow assessments of recent trends in the timing of the onset of childbearing for specific reference periods. The objective of this research note is to propose two different methods for the estimation of mean age at first birth from information collected in DHS and similar surveys. Both measures are unaffected by changes in the population age structure, thus allowing undistorted comparisons of the timing of the onset of childbearing between populations and over time within populations. Estimates are calculated for the most recent surveys in 62 countries.The equation for estimating the period mean age at first birth used widely in countries with vital statistics  is1\docuwhereM(t) = Average age at first birth at time tb = the age-specific birth rate for birth order one at (single) age a and time t.amax = the highest age at which first births are observedThis period mean age at first birth is defined as the mean age at which women would bear their first child if they went through the reproductive years having the first birth rates observed in a particular period.M(t) can be estimated.Numbers of births recorded in vital statistics are typically large and birth rates are available by single age and single year. As a result, annual estimates of b by single year of age for a period of three years before each survey. In addition, we exclude surveys with sample sizes of currently married women below 3000 to minimize sampling errors.In contrast, in applications of this equation to DHS surveys samples of births in a single year are relatively small. To obtain more robust estimates of the mean age at first birth for a survey, we calculate An alternative approach to estimating the period mean age at first birth is to rely on a method that is widely used to estimate the mean age at first marriage, called the “singulate mean age at marriage” , 12. TheM*(t) = Average age at first birth at time tp = Proportion of women that has not yet given birth at age a and time tpmax = The proportion of women that has never had a birth at amaxt.This mean age at first birth is defined as the mean age at which women would bear their first child if they went through the reproductive years experiencing the age-specific proportions childless observed at time M (t) = M* (t)) under the condition that the shape of the function p by age is invariant with respect to time. This implies that p can shift to high or lower ages over time (with corresponding changes in first birth rates and in the mean age) but with no change in shape and with constant pmax.In Additional file M(t) and M*(t) were obtained with equations (1) and (2) for the most recent DHS surveys in 62 developing countries for which data files are available for public use (and with sample sizes of married women above 3000).M(t) and M*(t) can also be calculated. Further details about the surveys are available on the DHS website [Estimates of quations  and (2)  website .b are obtained from birth histories with a simple variant of the standard DHS method for calculating age-specific birth rates by age for the three years before the survey [b two changes are made in this method: (1) birth rates are calculated by single year rather than by five year age intervals and (2) the numerators of the birth rates exclude births of order two and higher. Estimates of p are also calculated with a variant of the standard DHS method estimating the proportion nulliparous by single year of age rather than five year age intervals.Estimates of e survey . To estip are subject to substantial sampling errors at ages above 40, because the proportions childless at these ages are usually less than five percent and the number of respondents is smaller than at lower ages. To minimize the effects of these errors on estimates of the mean age at firth birth, the value of amax is set at 40 years and pmax is estimated as the average of single age values of p between ages 35 and 45.Finally, it should be noted that values of M(t) on the horizontal axis and the value of M*(t) on the vertical axis. Each marker represents the most recent survey in each of the 62 countries. The results are presented in two clusters: the solid markers represent surveys in which pmax is less than 5 % and the open circles represent surveys with pmax >5 %. This distinction is made to separate observations in which the conditions are met for M(t) to be equal to M*(t) from observations in which they are not. As noted in Additional file M(t) and M*(t) is that pmax is constant. Unfortunately, it is not easy to determine the rate of change in pmax, because some countries have only one survey and, even in countries with multiple surveys, the rate of change in pmax is erratic due to small sample sizes. Instead, we assume that countries with pmax less than 5 % have seen little change in pmax over time, thus approximating the condition that pmax is constant. In surveys where pmax is higher than 5 % there has likely been change over time because early in the fertility transition pmax is typically a very small number.Figure M(t) is closer to M*(t) for surveys in the cluster with pmax <5 %. As is evident from Fig. M(t) and M*(t) are respectively 21.0 and 21.2, a difference of only 0.2 year . However, the agreement is not perfect and the solid markers are spread around the diagonal in Fig. M*(t) is usually substantially higher than M(t). This finding is likely attributable to an upward bias in M*(t) when values of pmax are rising (the rare cases in this cluster with M(t) higher than M*(t) are probably attributable to measurement or reporting errors). Our working assumption therefore is that M(t) is an unbiased estimator of the mean age at first birth even in surveys in the second cluster. In addition, all except one of the countries in the second cluster have a mean age at birth of 22 or higher. This result is not unexpected as there tends to be a positive correlation between age at first birth and the proportion of women who remain childless.It is therefore expected that M(t) vary widely among countries from a low of 19.1 in Niger (2006) to a high of 24.7 in the Maldives (2009). The unweighted averages of M(t) for countries in each of four regions are presented in Table A full analysis of levels and trends in all 62 countries is beyond the scope of this methodological study, but a few findings can be noted. Estimates of M(t) for selected countries in the developing and developed world. Estimates for Egypt, Nigeria, India, Kenya, and Bangladesh show very modest increases from the 1990s to near 2010. The mean ages at first birth for the Japan, Czech Republic, UK, and US are mostly substantially higher and have been rising at a more rapid pace than in the five developing countries included in the figure.Figure As noted, DHS published reports provide estimates of the retrospectively reported cohort median age at first birth. These medians are estimated from birth histories obtained from respondents of reproductive age. The age at first birth is calculated by subtracting the woman’s date of birth from the date of birth of her first child. Medians for the cohorts aged 25–29 at the time of the survey and above are available for nearly all DHS surveys because the medians are reached before age 25 . For a small number of surveys medians are available for the cohorts aged 20–24 when the median is below age 20.These cohort medians have the advantage of being available for all DHS surveys but there are also drawbacks: 1) the median refers to past experience of cohorts and is therefore not as current as is preferable for many analytic purposes; 2) the retrospective reporting of the date of the first birth may suffer from recall errors that are likely to increase as the time since the event rises; and 3) the cohort median as calculated by DHS is not independent of the quantum of first births and can change over time even if the mean is constant. The first two of these disadvantages also apply to cohort mean ages at first birth, a measure we do not discuss because it is very rarely used as it can only be estimated accurately for women who have completed their childbearing.To illustrate, Fig. M(t) as a solid line based on five surveys between 1989 and 2008/9. (The points in this line are plotted 1.5 years before the survey date to account for the fact that the mean is based on births in a three year period before the survey.) The period mean shows a rise between the 1989 and 1998 surveys but remains flat from 1998 to 2008/9.Figure The period means and cohort medians are not directly comparable because they are different metrics of different distributions, but by plotting the data in comparable years (as discussed above) some tentative conclusions can be reached. In particular, the medians reported by women aged 25–29 are lower than the means. This pattern is as expected because the distribution of first births is skewed to higher ages. Comparisons of period means and cohort medians in other countries yield broadly similar results (data not shown).M(t) is based on recent births, the known biases in the reporting of distant first births by older women are likely to be minimized. Interestingly, our estimate of M(t) remains unaffected if errors are proportionally the same at all ages. The reason is that age-specific birth rates b appear in the numerator and the denominator of Equation 1. An error of say 10 % in all b values would lead to an error of 10 % in the TFR, but there would be no error in M(t). In reality errors in b are likely to vary somewhat by age and that would lead to a bias in M(t). Furthermore, errors in birth histories would not affect M*(t), unless women misreport their childlessness status at the time of the survey.It should be emphasized that survey data and any measures derived from them are subject to various reporting and non-reporting errors including omission of births, displacement of births in time, and variations in sample selection and implementation , 21–24. In addition to reporting errors in the birth history, the mean age at first birth estimates could be biased by women’s misreports of their own date of birth, especially if the misreporting is linked to fertility. If, for example, a woman who has begun childbearing early overstates her age due to negative social norms around early childbearing or if an interviewer estimates her age based on her childbearing status (in places where knowledge of birth dates is uncommon), then the mean age at birth would be overestimated. The completeness and accuracy of birth date reporting, of both women and their children, is likely to have improved over time, a factor that should be kept in mind when assessing trends.The timing of the onset of parenthood is a key indicator used in studies of the determinants and consequences of early childbearing as well as an indicator of the success of various programmatic interventions. Annual estimates of the period mean age at first birth from vital statistics are widely available in most developed countries. In contrast, vital statistics of high quality are lacking in the large majority of developing countries and sample surveys such as the DHS are the primary source of demographic and health indicators. The published indicators from these data include the retrospectively reported median but not the period mean age at first birth. Both medians and means are dependent on the quality of reporting in the birth history as well as reporting of their own birth dates by women.We assessed two methods to estimate the period mean age at first birth. The first method is the same as the one used in populations with accurate vital statistics, and the second is the singulate mean age at first birth. A comparison of the two estimates obtained from 62 DHS surveys shows excellent agreement in countries in which there is no evidence of an increase in childlessness. But, as expected on theoretical grounds, there is less agreement in populations that have experienced a rise in the proportion childless. We therefore prefer the first method. The measure is readily calculated as a straightforward variant of the standard procedure used by DHS to estimate period fertility rates and its reference period (the three years prior to the survey) is the same as the published total fertility rates. In addition, it refers to recent births and is, therefore, presumably more accurately reported than indicators based on events that occurred in the distant past. Since this new measure makes it possible for the first time to assess recent trends in the onset of childbearing in developing countries with multiple DHS surveys and to compare recent period estimates of the mean age at first birth among countries, we suggest that it be considered for inclusion in published DHS reports.


Dentigerous cyst is a development odontogenic cyst, whichapparently develops by accumulation of fluid betweenreduced enamel epithelium and the tooth crown of anunerupted tooth. It is one of the most prevalent types ofodontogenic cysts associated with an erupted, developingor impacted tooth, particularly the mandibular 3rd molars,the other teeth that are commonly affected are maxillarycanines and third molars. The present case report describesthe surgical enucleation of a dentigerous cyst involvingpermanent lateral incisor, the surgery was followed by oralrehabilitation. A cyst is defined as a pathological cavity lined by epithelium.The epithelium itself is surrounded by fibrocollagenousconnective tissue and may be derived from various sources.Odontogenic cysts are derived from the odontogenicepithelium which is derived from the basal epithelium ofthe stomodeum.3A nine year old male patient presented with a progressivelyenlarging painles swelling on the left side ofthe face over past 3 months  which caused elevationof the nasal floor and was also palpable on the palatalsurface.Clinical examination revealed a firm swelling fixed tothe alveolar process of the maxilla . Diagnosticmaxillary occlusal radiograph  and an orthopantomograph showed a radiolucent lesion in the alveolarprocess of the anterior maxilla. The lesion was well definedmeasuring approximately 3 × 2.5 cm. Lateral incisor wasassociated with the lining of the cyst and tooth was displaced.The contents of the swelling were aspirated and sent forinvestigation, the result of which was consistent with thediagnosis of cystic lesion. After clinical and radiologicalexamination a provisional diagnosis of dentigerous cyst wasmade. Prior to surgery, routine blood and urine examinationwere advised. The results were within normal limits. Surgicalenucleation of the cyst was chosen as the treatment of choice. The surgery was done using local anesthesiaand under antibiotic cover. The cyst was attached to thecementoenamal junction of left lateral maxillary permanentteeth. The specimen was sent for histopathological examination.The histological examination showed a thin fibrouscystic wall lined by a 2 to 3 layer thick nonkeratinizedstratified squamous epithelium with islands of odontogenicepithelium . The connective tissue showed a slightinflammatory cell infiltrate, which confirmed the diagnosisof dentigerous cyst. Following enucleation of the cyst, thepatient was recalled after 1 week for suture removal. As thelesion was big involving considerable amount of bone, leftcentral incisor became mobile after the surgery, hencefortha composite wire splint was placed . After 1 month,a removable partial denture was delivered, which servedfunctional space maintainer, improved esthetics andphonetics .Dentigerous cysts are the most common of the developmentalodontogenic cysts of the jaws and account for theapproximately 20-24% of all the epithelium lined jaw cysts.It develops around the crown of an unerupted tooth byexpansion of the follicle when fluid collects or a spaceoccurs between the reduced enamel epithelium and theenamel of impacted tooth. These cysts are often asymptomaticunless there is an acute inflammatory exacerbationand therefore these lesions are usually diagnosed duringroutine radiograph.Radiographic examination of a dentigerous cyst showsa well defined unilocular radiolucency often with a scleroticborder, surrounding the crown of an unreputed tooth.Histologically, the dentigerous cyst consists of a fibrous walllined by nonkeratinized stratified squamous epitheliumconsisting of myxoid tissue, odontogenic remnants andrarely sebaceous cells.Radicular cyst is the most common odontogenic cyst ofthe maxilla and mandible. Radiologically, it arises from theapex of the root of a carious tooth and is bounded by a thinrim of cortical bone. The differentiating feature of this entityis its relation to the root of a carious tooth.Odontogenic keratocyst results from cystic degenerationof the enamel organ before the tooth is formed so that thecyst replaces the tooth. It is commonly noted in the mandible.The classical feature of this cyst is the absence of the relatedtooth.Nonodontogenic cysts are observed in the region ofincisive canal or nasolabial regions. The incisive canal cystis in the midline located between the roots of the centralincisors of maxilla and is characteristically heart shaped.The nasolabial cyst occurs in the soft tissues of the lateralaspect of the nose and upper lip. These cysts are thereforediagnosed by their classical anatomical location. Aneurysmalbone cyst is seen as expansible multilocularradiolucent lesion. CT/MRI may reveal presence of bloodor fluid contents in the cyst.A dentigerous cyst associated with an anterior tooth willresult in failure or eruption of the tooth and therefore leadto esthetic and orthodontic problems. Absence of a lateralincisor can have an impact on the psychology of child.Further esthetic management has to be considered to preventand psychological trauma to the child. In the present case,esthetic management was done by providing the patient witha removable partial denture, which also serves as a functionalspace maintainer.


Multilevel random coefficient models were used to estimate the effects of PM2.5, SES and other individual and neighbourhood-level covariates on continuous birth weight and test interactions. Gestational age was modeled with a random slope to assess potential neighbourhood-level differences of its effect on birth weight and whether any between-neighbourhood variability can be explained by cross-level interactions.Birth records from 2001 to 2006 (2.5 and birth weight explaining 8.5 % of the between-neighbourhood differences in mean birth weight. A significant interaction between SES and PM2.5 was observed, revealing a more pronounced negative effect of PM2.5 on birth weight in lower SES neighbourhoods. Further positive and negative modification of the PM2.5 effect was observed with maternal smoking, maternal age, gestational diabetes, and suspected maternal drug or alcohol use. The random intercept variance indicating between-neighbourhood birth weight differences was reduced by 75 % in the final model, while the random slope variance for between-neighbourhood gestational age effects remained virtually unchanged.Models adjusted for individual and neighbourhood-level covariates showed a significant non-linear negative association between PM2.5 have both independent and interacting associations with birth weight, and together account for 49 % of the between-neighbourhood differences in birth weight. Evidence of effect modification of PM2.5 on birth weight across various maternal and neighbourhood-level factors suggests that certain sub-populations may be more or less vulnerable to relatively low doses PM2.5 exposure.We provide evidence that neighbourhood-level SES variables and PM PM2.5, which includes ultrafine particles less than 0.1 μm, can penetrate deep into the pulmonary alveolar tissue where inflammatory mediators and possibly the particles themselves translocate into the bloodstream causing systemic cardiovascular and immunological alterations such as platelet activation, coagulation and endothelial dysfunction [Studies of exposure to particulate air pollution have consistently shown an association with low birth weight, a predictor of fetal growth restriction and important determinant of infant and child-wellbeing –3. The ffunction –6. Thesefunction –9. Excesfunction .2.5 may act to adversely impact the reproductive system are not fully understood; however, evidence supports the potential for a shared mode of developmental toxicity with several other known risk factors [These mechanisms by which PM factors . This in factors , drug us factors , advance factors  gestatio factors , and low factors . The cau factors , 18. Thi factors , 20.A particular challenge in environmental epidemiology is handling data at differing geographic scales. Birth registries and vital statistics provide data on individual births and certain risk factors, but may not have data on socially patterned risk factors. Alternatively, reliable SES data such as education, income and housing quality are often only available from national census databases using arbitrary administrative spatial units. Finally, obtaining individual-level environmental exposure data is often not possible. Therefore, the epidemiologist is often left with a mix of individual-level observations clustered within neighbourhood areas each with distinct attributes. The use of multilevel statistical models separates the individual-level effects from the context of their social and physical environments and can therefore quantify the degree of clustering of individuals within neighbourhood areas and test whether neighbourhood factors themselves have direct effects on the health outcome or act indirectly via the modification of individual-level variables , 22.2.5-birth outcome relationship [2.5 in British Columbia, Canada where levels of PM2.5 are relatively low but can vary substantially between different communities [2.5, neighbourhood-level SES indicators, and/or individual-level risk factors are able to explain any neighbourhood-level variability. We had three research questions: 1) does exposure to PM2.5 and residence in low SES neighbourhoods in BC have significant independent negative associations with birth weight? 2) does the effect of gestational age on birth weight differ between neighbourhoods? 3) does PM2.5 interact with neighbourhood-level SES and/or individual-level risk factors to modify their independent effects on birth weight to help explain any neighbourhood-level differences?Through the mechanisms of oxidative stress and inflammation there is evidence that SES may not only confound but modify the PMtionship –25. Varimunities . We explN = 237,470). Data from the BC Perinatal Data Registry were provided by Perinatal Services British Columbia (PSBC) which included information on maternal-infant health status and outcomes, reproductive history, maternal risk factors and attributes, and residential postal codes. The Registry accounts for 99 % of births and stillbirths in BC of at least 20 weeks gestation or at least 500 g birth weight. Research data access is provided by a Partnership Accord /Memorandum of Agreement between all BC Health Authorities and PSBC through the Freedom of Information and Privacy Protection Act [This was a population-based retrospective cohort of singleton births in British Columbia from 2001 to 2006 (tion Act . Researcn = 1096), non-viable births prior to 20 weeks gestation or < 500 g (n = 14), and the list-wise deletion of births missing important data including: cigarettes smoked per day , PM2.5 (n = 1510), gestational age (n = 373), and birth weight (n = 41). All continuous variables, except cigarettes/day, were standardized and centred to ease interpretation and aid model convergence.The outcome variable was continuous birth weight of singleton births. In order to avoid potential selection bias, we included all births (stillbirth and live) for all gestational ages (20–44 weeks). Excluded records included out-of-province and invalid postal codes  model developed to estimate PM2.5 at the census street block-face level [2.5 concentrations. Unlike nitrogen dioxide (NO2), PM2.5 tends to have a more homogeneous intra-urban distribution between personal, indoor and ambient exposure [2.5 exposures between individuals within a given DA. We therefore aggregated the point-level estimates of PM2.5 to their DA-level mean and related it to individual birth records as an area-level variable.Exposure to PMce level . The modexposure . The LUR2.5 are also high. The correlation between immigrant density with SESi and PM2.5 was −0.62 and 0.53 respectively (p < 0.001); we therefore created a residual immigrant density variable using a sequential regression technique [2.5 with the saved residuals representing the uncorrelated and independent contribution of immigrant density on birth weight freed from its collinearity with SESi and PM2.5. This same method was used between SESi and education (r = 0.25) creating a residual education variable. The education and immigrant data were obtained by access to ABACUS via the Data Liberation Initiative [The DA-level SES and demographic data were represented by three related but independent datasets all based on the 2006 Statistics Canada national census. The first was a Canadian SES index (SESi) developed by Chan et al. . The secechnique . Here, iitiative .In order to avoid data loss from rural DAs, imputation for missing SES, education and immigrant density values was performed. Taking advantage of the nested hierarchical structure of the administrative census and health boundaries, the mean value for a larger encompassing census subdivision (CSD) or local health area (LHA) was imputed for a nested DA with a missing value. There were 1441 values imputed in 52 DAs for SESi , and 3170 values imputed in 108 DAs for both education and immigrant density . Sensitivity analyses were performed using only the non-missing data.R2 [Hierarchical (multilevel) linear regression models were used to test our research questions, thereby accounting for the clustering, or non-independence, of individuals (level-1) belonging to a given DA neighbourhood (level-2). The multilevel model allows the intercept and slope to act as random parameters having between-area (DA) variability from an overall (BC-wide) mean intercept and slope. That is, each DA has its own intercept and slope in which their variability from the overall intercept and slope can be investigated with the addition of level-1 and level-2 variables and their interactions . We follR2 . We starR2 .Stata 13IC [Gestational age was added to the null model and given a random slope . The presence of a significant random slope indicates that its effect is not constant  for all DAs. Subsequent models included the individual and DA-level variables along with cross-level and within-level interactions in order to assess their fixed effects on birth weight but to also determine if their inclusion addressed any unexplained intercept or slope variance. Models were tested using the Akaike Information Criterion (AIC) to evaluate model performance. All statistical analyses were conducted in ata 13IC .intra-area dependence while quantifying inter-area variance, they assume spatial independence among neighbouring areas. However environmental and social processes can extend beyond arbitrary neighbourhood boundaries. Additionally as mentioned above, census DAs do not necessarily represent neighbourhood dynamics, services, infrastructure, etc.; and evidence of spatial clustering between DAs may indicate that an alternative neighbourhood areal unit should be considered. We used spatial methods to test for this by checking the level-1 model residuals and level-2 predicted random parameters (intercepts and slopes) for spatial autocorrelation using the local Moran’s I statistic [Stata 13IC.Finally, while multilevel models address tatistic . The pretatistic  availabl2.5. Table 2.5 were also modeled using a quadratic terms, both indicating a subdued dose–response with increasing exposure.After exclusions there were 231,929 singleton (live and stillborn) births located in 6338 neighbourhood DAs . Table 2.5 and season of birth (cold vs. warm) in Model 3. Higher education had a positive association with birth weight, while rural areas had a significant negative association with birth weight. PM2.5 was found to have a significant non-linear negative association on birth weight whereby the negative effect tapers off at higher concentrations of PM2.5  and level-2 by level-2 interactions to explain the between-DA random intercept variability. The model results are presented in Tables 2.5. Four maternal variables showed effect modification with PM2.5 on birth weight. Maternal smoking and suspected drug or alcohol use both had positive interactions with PM2.5 on birth weight revealing a subdued association with increased PM2.5 exposure , increased to 2.3 %. This was due to the large reduction in the level-1 residual variance (560.5 to 435.2) relative to the reduction in the level-2 random intercept variance (78.8 to 67.0). The addition of DA-level variables in Model 2 and Model 3 removed a lot of the DA-level variance reducing the ICCadj to 1.1 and 0.8 % respectively.The random effects, the explained proportional change in variance (PVC), and model diagnostics are presented in Table The level-2 random intercept standard deviation indicates that the mean birth weight for every DA has a degree of variability from the overall (BC-wide) mean birth weight. For the Null model, the overall birth weight intercept is was 3434.2 g with a standard deviation of 78.8. This equals a 8.6 % difference in range between 95 % of the DAs (3434.2 ± (1.96 × 78.8) = 3280.0 and 3588.8 g). Similarly, we calculate the between-DA 95 % distributional range of slopes for gestational age to fall between 255.8 and 361.6 g (308.7 ± (1.96 × 26.6)), a 29.3 % difference in how one week gestation increases birth weight between DAs.2.5 and season of birth in Model 3 further explained an additional 8.5 % of the level-2 intercept variance beyond that of Model 2. Model 4 accounted for an additional 2.3 % L2-PCV.The level-1 and level-2 explained PCV (L1-PCV & L2-PCV) summarizes the relative degree of explained variance at the different levels between the different models. Using the Null model as the reference, the L1-model resulted in an L1-PCV & L2-PCV of 39.7 and 27.7 % respectively Table . These aSpatial analyses were used as a model diagnostic to test for significant spatial autocorrelation of the model residuals. The local Moran’s I statistics reported in Table 1 = 228,765 in 6230 DAs) showed very minor differences in magnitude of significant variables in the birth weight models. In a second sensitivity analysis, we restricted the sample to only term births excluding stillbirths and congenital anomalies. As expected, there was a large reduction in the random-slope variance for gestational age due to dropping preterm births but no other differences in the observed relationships . The interaction between gestational diabetes and PM2.5 was reduced by half, but was still significant. The effects of the DA-level variables SESi and immigrant density increased moderately beyond their 95 % CIs listed for Model 4. The effect for PM2.5 decreased but was still significant (-18.0 (95 % CI -21.3 – -14.7)). The DA-level interaction between PM2.5 and immigrant density was reduced by half and marginally not significant (p = 0.051), while the interaction between SESi and PM2.5 was also reduced to non-significance (p = 0.33).Finally, a check for potential collider bias was performed by omitting gestational age as a covariate from the models . Random-2.5 on birth weight and test its interaction with individual and neighbourhood-level risk factors. Our results show that individual and neighbourhood-level factors are capable of modifying the association between PM2.5 exposure and fetal growth. Furthermore, through the use of random-slopes models we show that the effect of gestational age on birth weight can vary considerably between neighbourhood DAs which was only moderately addressed in our models. After adjusting for individual-level covariates and DA-level socio-economic and socio-demographic variables, we found a significant non-linear effect between PM2.5 and birth weight in 231,929 births in British Columbia, Canada. This association was robust to the exclusion of stillbirths and congenital anomalies as well as the use of only term births and models dropping gestational age as a covariate, demonstrating that selection bias does not affect the observed PM2.5 main effects.This study employed multilevel random coefficient models to assess the effect of PM2.5 and birth weight [2.5 and is exemplified by Fig. 2.5, a not uncommon dose–response phenomenon also observed in other exposure-disease contexts [Our results corroborate the growing literature supporting a negative association between PMh weight , 46, 47.h weight . This stcontexts . Other scontexts , 50, 51.contexts .2.5 and SES such that a more pronounced effect of PM2.5 was seen in lower SES neighbourhoods  and may partially explain the observed rural–urban differences [2.5 was not statistically significant.The significant negative interaction between rural address and PMUR model . The comferences , 6. The ferences , 55. An 2.5 and maternal-level variables shown to reduce birth weight independently revealed some counter-intuitive results. This included the PM2.5 interaction with maternal smoking (cigarettes/day) and with suspected drug/alcohol use where increasing PM2.5 levels tempered the negative effect of these risk factors . The persistence of this finding leads to a hypothesis that some individual-level exposures may act as a pre-conditioning stress that activates an adaptive response of increased biological resistance to similar stressors [Interactions between PMors Fig. . This fitressors .2.5 exposure was also observed by Basu et al. [2.5 and other air pollutants [2.5. This significant negative interaction between PM2.5 and gestational diabetes could be related to excess of systemic or placental oxidative stress and inflammation resulting in restricted fetal growth [A protective effect of older maternal age against PMu et al. , and mayu et al. , 58. Curu et al. . Gestatillutants , 60; howl growth , 61.While the application of multilevel models in perinatal epidemiology has become more common , most haSpatial analyses were used to examine the wider spatial context within which the DAs are situated and also served as a measure of model specification. The inclusion of the DA-level variables and interactions substantially reduced the spatial autocorrelation in the level-1 and level-2 random-intercept residuals . We therefore assume all pregnancies were exposed to the same levels of PM2.5 for their entire pregnancy, regardless of their year of birth, based on their residential DA. While this method prevents the assessment of exposure windows by trimester, spatiotemporal studies of PM2.5 have shown little to no difference between trimester-specific and entire pregnancy effects on birth weight [2.5 concentrations may be underestimated by the LUR model with less variability and missing several high PM2.5 outlier locations in BC compared to compiled monitored data [2.5 levels.A key component of this research was the use of a land-use regression (LUR) model of air pollution . While th weight , 46, 48.red data . This co2.5 effect on birth weight could be due to individual-level SES factors not accounted for in our models such as diet, income or stress. However, studies have found that adjustment for individual-level measures of SES did not significantly change the area-level associations [We were unable to control for maternal-level SES, and therefore the neighbourhood-level effect estimates and interactions could reflect individual-level differences. For example, the protective effect of older maternal age buffering the PMciations , 65. Matciations .2.5 on birth weight and its modification by both maternal and neighbourhood-level factors. Most notably, it shows that lower SES neighbourhoods may be more negatively affected by higher levels of PM2.5. We observed both positive and negative interactions between maternal factors and PM2.5 that require further scrutiny but may reflect a PM2.5-oxidiative stress pathway expressed via either protective pre-conditioning or harmful overload. Targeted municipal-level interventions to reduce PM2.5 and improved neighbourhood SES may help improve birth outcomes at the population-level.This study supports the growing literature of an effect of PM


We delved into the selective migration hypothesis on health by comparing birth outcomes of Latin American immigrants giving birth in two receiving countries with dissimilar immigration admission policies: Canada and Spain. We hypothesized that a stronger immigrant selection in Canada will reflect more favourable outcomes among Latin Americans giving birth in Canada than among their counterparts giving birth in Spain.We conducted a cross-sectional bi-national comparative study. We analyzed birth data of singleton infants born in Canada (2000–2005)  and Spain (1998–2007)  to mothers born in Spanish-speaking Latin American countries. We compared mean birthweight at 37–41 weeks gestation, and low birthweight and preterm birth rates between Latin American immigrants to Canada vs. Spain. Regression analysis for aggregate data was used to obtain Odds Ratios and Mean birthweight differences adjusted for infant sex, maternal age, parity, marital status, and father born in same source country.Latin American women in Canada had heavier newborns than their same-country counterparts giving birth in Spain, overall , and within each maternal country of origin. Latin American women in Canada had fewer low birthweight and preterm infants than those giving birth in Spain .Latin American immigrant women had better birth outcomes in Canada than in Spain, suggesting a more selective migration in Canada than in Spain. The healthy migrant effect  refers to the phenomenon wherein immigrants are healthier than the local native-born population and their co-nationals who did not emigrate. The healThe selective migration hypothesis has been examined most extensively in the context of adult health, and mostly among immigrants in the United States. Some of these studies found some support for the selective migration hypothesis–6 but otImmigration of Latin Americans to Canada and Spain provides an opportunity to study the influence of selective migration on health. Indeed, for many Latin Americans, immigrating to Canada has been more restricted than to Spain. As a traditional immigrant-receiving country Canada’s immigration regime has evolved since the country’s founding and currently most immigrants to Canada are selected on the basis of a point system that rewards characteristics associated with success in the labor market, such as working knowledge of English/French as a second language, education, work experience and wealth. Canada also admits lower human capital immigrants such as refugees and family-class migrants to a lesser extent. Unauthorth century.[Latin Americans face fewer challenges to migrate to Spain than to Canada because of the common Spanish language and citizenship rights among those who have a Spanish ancestor up to two generations. Latin America has been the main destination of Spanish emigrants from the sixteenth century to the 1970’s. About 2.5 million Spanish emigrants headed towards Latin American countries during the 20 century. By 2006,Based on the marked differences between the admission contexts of the two countries, if selective migration is associated with better health we would expect more favourable outcomes among Latin Americans who migrated to Canada compared to those who migrated to Spain, as a result of Canada’s stricter selection and admission policies. Low birthweight and preterm birth are adverse pregnancy outcomes sensitive to environmental influences, such as social class. They areWe delved into the selective migration hypothesis by comparing mean birthweight at term, low birthweight and preterm birth between Latin American immigrants giving birth in Canada vs. Spain, using comparable contemporary nationwide data.This cross-sectional population-based comparative study uses national birth certificate data from Canada and Spain. The study population is composed of singleton liveborn infants in Canada (2000–2005) and Spain (1998–2007) to women born in Spanish-speaking Latin American countries and native-born women (Canadian-born women in Canada and Spanish-born women in Spain) in the last decade available in national birth certificate data. The longer study period for Spain reflects availability of data; however, the midpoint is the same for both countries, thus maintaining comparability. Both registers collect information on maternal country of birth and other sociodemographic characteristics. To minimize bias due to potential outliers, we only included records with birthweights within 250–5999 grams and gestational age within 20–45 weeks.Information on birth outcomes in Canada was obtained from the linked live birth-death file of the Public Health Agency of Canada’s Canadian Perinatal Surveillance System. Approval for use of this data was obtained from the Agency and Statistics Canada. Ethics approval was not legally required since data were anonymized, analyses were conducted in restricted access Research Data Centers, and guidelines for release of results were adhered to. Likewise, this study only used anonymized Spanish birth certificate data compiled by the National Institute of Statistics and the review of an Ethics Research Committee was not legally required in Spain.Secondary analyses were conducted with a subsample of Latin American women delivering in Ontario hospitals (1998–2007) and captured through a linkage between the official immigration register, the Citizenship and Immigration Canada database, and the Discharge Abstracts Database, which comprise all hospital births. Data collection methods and use of Ontario hospital data were approved by the St. Michael’s Hospital Research Ethics Board. These analyses were conducted at the Institute for Clinical Evaluative Sciences (ICES) in Toronto, Ontario. ICES is a prescribed entity under Section 45 of the Ontario Personal Health Information Protection Act (PHIPA) 2004. As such, it is permitted to receive (collect) and use personal health information without consent. ICES implements strict privacy and confidentiality practices and procedures that have been reviewed and approved by the Information and Privacy Commissioner of Ontario as required under the Act.Outcome measures were singleton mean birthweight at term (37–41 weeks gestation), low birthweight (< 2500 grams) and preterm birth (<37 weeks of gestation). In Canadian birth certificates, the information on gestational age is recorded by the attending physician, nurse or midwife. In Spain, it is recorded by the parents based on the information provided to them by the birth attendant. A Spanish validation study using records of a single hospital as the gold standard, found that the sensitivity and specificity of birth certificate data for low birthweight were 90% and 99%, respectively and 78% and 98% for preterm birth. Most notMaternal country of birth included Canada, Spain, Argentina, Bolivia, Chile, Colombia, Cuba, Ecuador, El Salvador, Honduras, Mexico, Nicaragua, Paraguay, Peru, Uruguay, Venezuela and Rest of Latin America . The latter category was created by collapsing countries contributing very few births in at least one of the two countries. Country of delivery included Canada and Spain.For descriptive statistics, we conducted a side-by-side analysis based on common definitions of the outcomes and birth characteristics in the two national datasets. Due to the restricted access of each dataset, it was not possible to pool the datasets of the two countries at the individual level but we could do so in aggregate form. In orderFinally, we conducted secondary analyses restricted to Ontario hospital data linked to the official immigration register (Citizenship and Immigration Canada Database) to assess whether time since arrival to Canada was associated with improved birth outcomes of Latin Americans in Ontario, as a potential alternative explanation to our findings.The share of singleton births to Latin American immigrant women was larger in Spain (4.56%) than in Canada (2.18%) . Latin AThe mean birthweight at term was significantly higher for infants born to Latin American women residing in Canada compared to their counterparts residing in Spain, before  and after  adjustment . FurtherThe rates of low birthweight and preterm birth were lower among infants born to Latin American women in Canada compared to those born to Latin American women in Spain . The adjSecondary analyses using the linked Ontario hospital birth records to the Canadian immigration register showed that time from arrival to delivery was not associated with increases in mean birthweight at term . Low birOur findings provide support to the hypothesis that selective migration may influence the more favourable birth outcomes of Latin American women delivering in Canada vs. in Spain. First, the share of singleton births to Latin American immigrant women was larger in Spain (4.56%) than in Canada (2.18%), despite the physical proximity of Latin America to Canada. Second, compared to Latin American women giving birth in Canada, those delivering in Spain were more likely to be teenagers, unmarried and not having delivered previously; these being correlates of adverse outcomes and factors associated with negative selection. Adjustment for these factors reduced the differences but did not eliminate them. Third and more important, birth outcomes were more favourable among Latin American women delivering in Canada vs. those delivering in Spain. Notably, mean birthweight at term was 59 to 212 grams higher among infants born to foreign-born women from different Latin American countries giving birth in Canada compared to those of their same-country counterparts giving birth in Spain. The magnitude of these differences is comparable to the effect of smoking on birthweight.,30Strengths of our study include the use of birth certificate data at the national level, comparable study periods and similar definition of the outcomes, maternal country of birth and covariates for statistical adjustment. Restriction of the study to a distinct immigrant group in two different receiving countries may diminish confounding resulting from unmeasured factors, since for example Mexican-born women giving birth in Canada may share background and cultural characteristics with their Mexican counterparts giving birth in Spain.Limitations are several. First, gestational age information in Spanish birth certificates may slightly overestimate preterm birth rates, and consequently, low birthweight rates, according to a validation study that linked about 9000 birth certificates with records of one Madrilean hospital. If this While low birthweight and preterm birth mainly reflect pathological processes interrupting the pregnancy or its normal development, mean birthweight at term is a measure of normal fetal growth in the population. If thereSecond, although we controlled for some maternal characteristics associated with birth outcomes, birth certificates do not provide information on other potentially relevant information, such as pre-migration characteristics , health behaviors during pregnancy, and post-migration social position and health care use. To compensate, we further adjusted for comparability scores designedThird, the lack of information on the type of migrants and the motivations for migration hampers our ability to explore in greater depth the nature of selective migration in our main analyses. Our a priori hypothesis assumed a pattern based on legal and voluntary migration. While most immigrants fall within this group, many immigrants may also be refugees or unauthorized. Births to both refugees and unauthorized immigrant mothers are captured, however, in both Canadian and Spanish birth certificates but it was not possible to identify them to conduct sub-group analyses. In the Ontario sub-sample, however, information on refugee status was available but refugee status was not associated with the study outcomes. The Ontario cohort was restricted to legal immigrants to Canada and therefore could not shed light on whether birth outcomes are different between legal and undocumented migrants.Finally, we could not provide a more complete picture of the potential effects of selective migration on birth outcomes by including data of Latin American women who did not migrate and gave birth in their home countries.To our knowledge, this is the first study focusing on the impact of selective migration on birth outcomes in relation to the immigrant admission context of the receiving countries. A recent study comparing preterm birth rates of Mexico-born women delivering in Mexico and in California did not find evidence of a healthy immigrant effect. This stuOur findings suggest that characteristics of the immigrant-receiving countries may be associated with differential selective migration patterns, which in turn may translate into different health outcomes. More specifically, the more restrictive immigration admission context in Canada seems to produce a “cream skimming” effect, further selecting women with characteristics conducive to favourable birth outcomes beyond self-selection to migrate. Selective migration of Latin Americans may take place both in Spain and Canada, but our findings suggest that it is stronger in Canada. While self-selection to emigrate may result in better birth outcomes among those who emigrated compared to those who did not migrate, Canada’s more restricted admission policies place a second filter on those who self-select themselves for migration, thus boosting the selection effect.In addition, Spain’s Latin American immigrant population has grown since the 2000s partly because of high demand for low-skill temporary laborers, many of whom became permanent residents through the 2005 regularization program. In Canada, there has not been a similar strong pull factor attracting lower SES Latin American immigrants.While the selection hypothesis places emphasis on pre-migration characteristics, a potential alternative explanation to our findings is that the observed differences may reflect a more favourable post-migration socioeconomic environment in Canada than in Spain. While most Latin American women giving birth in Spain were recent immigrants, since immigration to Spain is a recent phenomenon that achieved its peak during the study period, immigratDifferences in the proportion of refugees in Canada and in Spain are unlikely to contribute to explain the observed differences. In our sensitivity analyses restricted to Ontario data, we did not find an association between refugee status and birth outcomes. Refugees to Spain are very few and unliFinally, the presence of unauthorized immigrants may play an important role in explaining our findings. Although the unauthorized immigrant population in Canada is small, it constIn sum, our findings provide support for the selective migration hypothesis, particularly due to a stricter immigrant selection in Canada compared to Spain. While most research on immigrant health focuses on the characteristics of the immigrant populations in relation to their places of origin, this study highlights the importance of paying attention to the destination environments, particularly the combination of “push-pull’ factors may determine the health trajectories of newly arrived immigrants. Future studies focusing on the differences in health outcomes between authorized and unauthorized migrants may help clarify the roles of self- and state-selection and the healthy migrant effect.