The rate of macrosomia  increased over the past four decades in many parts of the world. Macrosomia is associated not only with higher risks of maternal and neonatal complications but also with health risks in adulthood. We examined trends in neonatal macrosomia and large-for-gestational-age (LGA) births among singleton, live, term and postterm births (≥37 complete weeks' gestation) in southeast China from 1994 to 2005 and explored possible causes of the temporal trends.Data from Perinatal Health Care Surveillance System in 12 cities and counties in southeast China were analyzed for trends in birth weight, neonatal macrosomia and LGA from 1994 to 2005. A total of 594, 472 singleton live births were included. We conducted multiple logistic regression analyses to relate these trends to changes in maternal and pregnancy characteristics.The rate of macrosomia rose from 6.00% in 1994 to 8.49% in 2000 and then levelled off to 7.83% in 2005. Similar trends were observed in mean birth weight. The incidence of LGA births increased continuously from 13.72% in 1994 to 18.08% in 2000, but the LGA rate remained relatively stable from 2002 to 2005. There was a decrease in gestational age and a significant increase in frequency of prelabor caesarean delivery from 1994 to 2005. In an adjusted multivariable model, the increase in LGA rate from 1994 to 2000 was associated with increasing net gestational weight gain, maternal age, maternal height and maternal education. But they didn't fully explain the increase. The trends of 2002-2005 LGA declined after adjusted for maternal and neonatal characteristics.In southeast China, the incidence of macrosomia increased from 1994 to 2000 was mainly related to increasing net gestational weight gain. The incidence of macrosomia has levelled off in recent years partly due to increasing use of prelabor caesarean delivery and earlier delivery and partly due to moderation of gestational weight gain. Birth weight and rate of macrosomia and large for gestational age (LGA) increased over the past four decades in many countries -7. TempoHowever, a recent study showed that the rate of macrosomia was decreasing in the United States since early 1990s  despite With rapid economic growth in China in the past three decades, investments in education, healthcare and sanitation have increased accordingly. Chinese national health services survey showed that birth weight increased from 3186 g in 1993 to 3284 g in 1998 and to 3307 g in 2003 . A rapidWe used data from the population-based PHCSS that was established along with the community intervention trial for preventing neural tube defects with peri-conceptive supplementation of 400 μg folic acid daily. The intervention trial was conducted in one northern province (Hebei) and two southeast provinces (Zhejiang and Jiangsu) from October 1993 through December, 1996. The original project design and results of the neural tube defects prevention program have already been published elsewhere [In the PHCSS, we collected information on parental demographics, maternal medical and reproductive history, medical conditions during pregnancy, labour and delivery summary and postpartum conditions. Each woman was issued a Perinatal Health Care Booklet and was assigned a unique identification number when she was at her marital registration, at the first prenatal care visit, or when she sought perinatal care services. The women were followed during their pregnancy, delivery and the immediate postpartum period by local health care professionals. After postpartum discharge, the booklet was collected and all data recorded in the booklet were computerized by trained staff at each county using a standardized data entry application with built-in data checking function. The National Center for Maternal and Infant Health was in charge of data cleaning and editing. In 2001, a computer electronic record system was established in the southeast provinces. Information on each prenatal visit, labor and delivery summary, and postpartum visit was entered by the health care providers on site and uploaded to the county server each day. The county then sent the data electronically every month to the National Center for Maternal and Infant Health for cleaning and editing.We selected 12 cities and counties from 19 surveillance sites because they had better data quality. They had the most complete data with, on average, over 90% of the births in the site, except for 2001. In 2001 problems with transitioning from paper to electronic registration resulted in inadequate data on large numbers infants. Thus, we dropped the 2001 data from the study. The data in this paper goes from 1994 through 2005 with the exception of 2001. The number of infants with missing birth weight was no more than 2.0% at all sites. To identify and delete implausible birthweight-gestational age combinations we used Alexander's algorithm [There were 692, 330 live births born from January 1, 1994 to December 31, 2005 in our study sites. We excluded 21, 914 births with missing birth weight, or gestational age values outside the range of 20-44 weeks or implausible birthweight-gestational age combinations. After exclusion of 40, 506 multiple births and preterm newborns, there were 629, 910 live-born singletons≥37 gestational weeks. Not all births were captured in 2001, we, therefore, deleted 2001 data, leaving 594, 472 births for analysis.2(Under weight), 18.5-22.9 kg/m2, 23-24.9 kg/m2(Overweight), ≥25 kg/m2(Obesity). Gestational weight gain was defined as the last measured weight in the 3rd trimester minus the early pregnancy weight or weight measured at premarital consultation, referring to IOM [Birth weight was recorded in 50 grams. Macrosomia was defined as birth weight ≥4, 000 g, irrespective of gestational age . Gestati2 tests for linear trend were used for dichotomous outcomes, and one-way analyses of variance with linear contrast were used for continuous outcomes. We used logistic regression to examine the associations between maternal and perinatal characteristics and risk of delivering macrosomia. We also estimated a series of multiple logistic regression models to examine the changes of single-year effects on LGA after sequentially adjusting for related factors. Values for maternal early pregnancy BMI and net gestational weight gain were missing for a substantial proportion of the study cohort. We assessed the impact of missing values by creating a category of "unknown" for factors with a substantial proportion of missing values or by deleting the cases with missing values. We then compared the results. The direction of the results didn't alter. Thus, the results presented here were the model without missing values. The statistical analyses were conducted with SPSS 11.5 .χOver the 11-year period, mean birth weight for all term and postterm infants increased from 3296 g in 1994 to 3378 g in 2000, then levelled off to 3369 g in 2005. As show in Figure The trend of incidence of macrosomia followed the pattern of birth weight for all term and postterm infants. The proportion of macrosomia increased steadily from 6.00% at 1994 to 8.49% at 2000 and moderated to 7.83% at 2005 (Figure The mode of delivery changed greatly in this period at our study sites. Figure Table To investigate the association between macrosomia and the potential explanatory variables among infants born at ≥37 weeks of gestation, we computed unadjusted and adjusted ORs with 95% confidence intervals Table . The matSince the incidence of macrosomia is affected by the duration of gestation, we also examined the trend of LGA, which better reflects fetal growth. We further separated the whole data into two periods: 1994-2000 and 2002-2005, because these two periods had different velocities of weight increase. The crude and adjusted odds ratios associated with yearly changes in LGA are showed in Table Our population-based data show that the mean birth weight, proportion of macrosomia and LGA increased overall in southeastern China from 1994 to 2005. The most important risk factor for the increase in macrosomia rate is net gestational weight gain. The plateau of macrosomia rate in recent years may be caused by declining fetal growth as a result of efforts to control net gestational weight gain as well as earlier delivery. This observation was consistent with findings from a hospital-based study from 1981 to 2006 .Changes in mean birth weight and incidence of macrosomia appear to have gone through two distinct periods: a rapid rise from 1994 to 2000 and a relative plateau from 2002 to 2005. This is parallel to the prelabor caesarean delivery rate: between 1994 and 2000 the rate of prelabor caesarean delivery rose from 20% in 1994 to 40% in 2000, and to 53% by 2002. It remained relatively stable from 2002 through 2005. The jump in rate of prelabor caesarean delivery was associated with a decrease in gestational age, and more macrosomic infants delivered by caesarean delivery. The increase in prelabor caesarean delivery may have been due to concerns about related complications for larger infants prompting delivery at an earlier gestational age. This finding corroborates observations reported by Schack-Nielsen , using DThe effect of gestational weight gain on birth weight may differ according to maternal prepregnant BMI. Most studies suggest that pregnancy weight gain influences birth weight more in women who are underweight than in women who are overweight ,28. ChinThe changing trends of macrosomia rate were associated with the dietary structure. In the 1990's, with economic development and improvement of living standard, nutrition and diet during pregnancy also improved. The dietary structure in Chinese population changed towards high calorie, high fat and low fibre. Women gained more weight and fetus grew faster. Net gestational weight gain reached its peak in 2000. This was temporally related to a concerted effort begun in 2001 to have the mother gain only an appropriate amount of weight during pregnancy. From 2002 through 2005 the net gestational weight gain plateaued, and so did the rate of macrosomic infants.The strength of our study was the use of surveillance data that provides a large population size to allow examination of small yet consistent changes in macrosomia over time. Unlike smaller and hospital-base studies, the population-based study permits us to generalize the findings to substantially larger areas of China. Second, our study area is one of the most developed regions in China. This is reflected in the great improvement in women's education. Thanks to economic reforms China has undertaken over the past few decades, many regions have substantially improved living conditions. Therefore, the trends observed in our study are instructive for the rest of China.Limitations of our data should also be noted. Our surveillance data did not record gestational age based on ultrasound dating. Gestational age based on the first date of last menstrual period has errors particularly among preterm and postterm births . Second,In southeast China, the incidence of macrosomia increased from 1994 to 2000 and was mainly related to increasing net gestation weight gain. The incidence of macrosomia has levelled off in recent years partly due to increasing use of prelabor caesarean delivery and earlier delivery and partly due to moderation of gestational weight gain.The authors declare that they have no competing interests.YL, JZ, and ZL were responsible for the conception, design and acquisition of data. YL was responsible for the analysis and interpretation of data and drafting the initial manuscript. JZ revised it critically for important intellectual content. XL and WX helped for the data analysis and modification. ZL were responsible for reviewing all drafts of the manuscript and giving final approval of the version to be published. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2458/11/818/prepub


Researchers and clinicians have for several decades focused epidemiologic, basic biomedical and clinical investigation on preterm birth, low birth weight, and numerous sequelae of untimely birth. As the problem has been examined from the expertise of various specialists and from separate but at times overlapping disciplines, different etiologic elements have emerged as important in the genesis of prematurity. Despite continuing discoveries in specific disciplines, preterm birth has not been etiologically tied to one factor such as genetic polymorphism, microbial challenge, or immune dysregulation. In recent years, the idea of “complex disease” has been proposed as a conceptual model for conditions that involve interplay between genetic, environmental, and other factors. This seems an apt description for preterm birth which will explain the breadth of topics in this special issue.This special issue begins by considering microbial challenges to successful pregnancy. Certainly, the concept of infectious disease as a factor in preterm birth and its sequelae must consider the microorganisms that inhabit the normal host and the way in which host immune system interacts with these microbes. The first five contributions to this issue are focused primarily on microbes that can threaten the pregnancy. Foremost in understanding the microbial challenge to pregnancy is having the tools to appropriately characterize the maternal microbiome. In the paper by X. Zhou of Forney's laboratory, we are reminded that only since new, culture-independent, DNA-based methods of analysis have emerged we are able to correctly characterize the lower genital flora and establish that there is no longer one single list of microbes that can be called “normal.” These findings provide important clarity to concepts regarding flora and preterm birth. Streptococcus (GBS). The concern for early identification and therapeutic intervention for GBS colonization in pregnancy has resulted in protocols that have become widely established. But experience over the past few decades has allowed a retrospective look at protocols again in specific clinical settings.The second and third contributions in this series return to the problem represented by Group B S. Faro et al. explore the experience of a private American hospital with respect to GBS screening and treatment. In 2 years of the study, they had an 89 percent screening rate with a 29 percent culture-positive rate with an invasive GBS incidence of 0.94 cases per 1000 live births. This is useful in underscoring the continuing need for vigilance for GBS. The third paper in this issue, also on the topic of GBS, looks at the issue with neonates exposed or presumptively exposed to the organism. B. Buckler and coworkers suggest that some workup of asymptomatic neonates “inadequately treated intrapartum” may be more costly and no better than clinical observation and treatment as warranted. Clearly, clinicians continue to seek refinements to their approaches to individual pathogens, and while the approach to GBS may be seen by some as firmly established, some details of the approach to this organism in pregnancy are open to continued review and assessment. Bacterial vaginosis is a topic that always emerges in discussions of preterm birth because of its epidemiologic association and the expansive literature that deals with the bacteriology and mechanistic associations with adverse pregnancy outcomes. Interest continues in finding antimicrobial approaches to the condition which in the case of the paper by I.-M. Bohbot and colleagues was aimed at clinical evaluation of secnidazole which they tested head to head with metronidazole. Mycoplasma and Ureaplasma, are frequently mentioned as having relationships to adverse pregnancy outcome, but in reality, these organisms are rarely cultured routinely in clinical practice. J. Hwang and B. Larsen have provided an update on the current understanding of these organisms in pregnancy and have reviewed research that reveals the important role of these microbes in eliciting cytokine responses that can be contributors to preterm birth.The fifth paper continues the description of specific organisms of concern in infectious conditions that may threaten pregnancy. The cell-wall deficient bacteria, Mycoplasma and Ureaplasma provide a nice segue into the sixth paper by J. E. Thaxton and coworkers. A cogent and timely summary of some of the important immune response mediators that can influence premature labor and delivery is provided as the conceptual connection between microorganisms and the immune responses that can lead to preterm birth. More specifically, this report provides a compelling account of the role of toll-like receptor engagement with intrauterine bacteria and the possibility of a cytokine storm involved in prematurity. The possible existence of a second uterine inflammatory pathway is an additional intriguing concept raised by this paper.The cytokine responses to As a continuum from term birth to late preterm birth, early preterm birth is one that may consider the most significant result of infectious complications of pregnancy to be the loss of the baby, and pregnancy loss is a problem of international importance with some 4.5 million stillbirths occurring annually worldwide. Thus, it is fitting that this issue includes a paper that recognizes this problem and offers some hopeful information. A cross-sectional study of plasmodial and helminthic infections in Ghana is reported by N. Yatich and coworkers who found a higher risk of stillbirth among pregnant women with parasitic infection and suggested that some pregnancies might be salvaged with early intervention if these conditions can be recognized earlier. The final paper in this issue by F. Omeaca et al. relates to the issues facing the preterm infant and one of the important threats to its survival and subsequent well-being. The authors explore the immune response of premature infants to hepatitis B vaccine as compared to term babies. In this report, while 6 of 94 premature infants did not respond, there were no significant differences in overall response to immunization due to low birthweight or gestational age suggesting the ability for vaccine application in this population.The editors of this issue are pleased to offer paper that are not artificially focused but rather through the breadth of topics underscore the concept of complex disease mentioned at the beginning of this editorial. The complexity of genetics, immunity, and microbes acting in a tangled web of interactions reminds the reader why the issue of preterm birth has proven so intractable and why many different disciplines have something to contribute to the ultimate understanding of this clinical problem. But we are hopeful that as the reader considers these varied contributions, it will be appreciated that despite the clinical descriptions and scientific discoveries, there is an underlying fact that deserves remembering. The clinical and scientific verities involved in adverse pregnancy outcomes are very personally felt by families that lose babies to infection or who have to deal with infants who are severely compromised by their prematurity. These parents undoubtedly are hopeful that the pursuits of scientists and physicians will move ever closer to practical answers to this very prevalent and vexing problem that intrudes on the happiness that is otherwise part of the birth of a child. Bryan LarsenBryan LarsenFrancesco DeSetaFrancesco DeSetaJoseph HwangJoseph HwangMario MerialdiMario MerialdiJosé Tirán-SaucedoJosé Tirán-Saucedo


Background: Air pollution exposure during pregnancy might have trimester-specific effects on fetal growth.Objective: We prospectively evaluated the associations of maternal air pollution exposure with fetal growth characteristics and adverse birth outcomes in 7,772 subjects in the Netherlands.10) and nitrogen dioxide (NO2) levels were estimated using dispersion modeling at the home address. Fetal head circumference, length, and weight were estimated in each trimester by ultrasound. Information on birth outcomes was obtained from medical records.Methods: Particulate matter with an aerodynamic diameter < 10 μm : –0.24, –0.12 mm; and –0.12 mm, 95% CI: –0.17, –0.06 mm per 1-μg/m3 increase in PM10 and NO2, respectively]. Average PM10 and NO2 levels during pregnancy were not associated with head circumference and length at birth or neonatally, but were inversely associated with birth weight . Longitudinal analyses showed similar patterns for head circumference and weight, but no associations with length. The third and fourth quartiles of PM10 exposure were associated with preterm birth . The third quartile of PM10 exposure, but not the fourth, was associated with small size for gestational age at birth (SGA) . No consistent associations were observed for NO2 levels and adverse birth outcomes.Results: In cross-sectional analyses, NO10 exposure was positively associated with preterm birth and SGA.Conclusions: Results suggest that maternal air pollution exposure is inversely associated with fetal growth during the second and third trimester and with weight at birth. PM Maternal exposure to air pollution during pregnancy has been suggested to be associated with increased risks of adverse birth outcomes such as low birth weight, intrauterine growth restriction, and preterm birth . Thus fa10) and nitrogen dioxide (NO2) exposure levels during pregnancy with fetal growth characteristics assessed by ultrasound in each trimester of pregnancy and adverse birth outcomes in a population-based cohort study among 7,772 pregnant women in the Netherlands.We investigated associations of particulate matter with an aerodynamic diameter < 10 μm , abortion (n = 7), or intrauterine death (n = 12) were excluded. The cohort for analysis consisted of 7,772 mothers and singleton live births .herlands . MothersAir pollution exposure. Individual exposures to PM10 and NO2 during pregnancy were assessed at the home address, using a combination of continuous monitoring data and dispersion modeling techniques, taking into account both the spatial and temporal variation in air pollution. A detailed description and flow chart of the exposure assessment are presented in Supplemental Material, pp. 3–6 and Figure S2 (http://dx.doi.org/10.1289/ehp.1003316). In brief, annual average concentrations of PM10 and NO2 for 2001–2006 were assessed for all addresses in the study area, using the three Dutch national standard methods for air quality modeling , taking into account hourly wind conditions and fixed temporal patterns in source contributions. We obtained full residential history of the participants. Based on the home addresses of the participants, we derived individual exposure estimates for different periods in pregnancy: conception until first-trimester ultrasound, conception until second-trimester ultrasound, conception until early third-trimester ultrasound, and conception until delivery.modeling . SubsequFetal growth characteristics. Fetal ultrasound examinations were performed in each trimester of pregnancy. In the first trimester, we used fetal crown–rump length to assess fetal growth only in mothers with a known date of last menstrual period (LMP) and a regular menstrual cycle of 28  days (32) days . Date of32) days , because32) days . Fetal g32) days . Longitu32) days .Birth outcomes. Medical records and hospital registries were used to obtain information about date of birth, gestational age at birth, fetal sex, birth weight, and birth length. We completed missing data on birth length (16%) with measurements of length from the first visit at the routine child health center within the first 2 months after birth, which had negligible influence on the results. Head circumference was not routinely measured at birth; therefore, we used head circumference from the first child health center visit. The regression models with neonatal length or head circumference as the outcome were adjusted for postconceptional age . Models with neonatal length were also adjusted for the method of measurement . Gestational age and sex-adjusted SD scores for birth weight and birth length were constructed based on reference charts from a North European birth cohort , parity , folic acid supplementation use  (denL) at the home address was assessed according to requirements of the European Environmental Noise Directive (http://dx.doi.org/10.1289/ehp.1003316).y, none) , and ethirective . The assStatistical analysis. We used the lowest quartiles of PM10 and NO2 exposure as the reference exposure groups. First, with multivariate linear regression models, we assessed associations between air pollution exposure in quartiles in the relevant time periods  with absolute measures of fetal growth and neonatal parameters. Second, to assess potential nonlinear longitudinal effects, we used mixed-effects models with unstructured residual covariance to longitudinally model fetal growth SD scores from 18 weeks of pregnancy until birth by natural cubic splines  and for road traffic noise exposure. Models of absolute measures of fetal growth, fetal growth SD scores, preterm birth, and low birth weight were additionally adjusted for fetal sex. Models of absolute measures of fetal growth were additionally adjusted for gestational age at measurement. Models of low birth weight were additionally adjusted for gestational age at birth. The percentages of missing values within the population for analysis were < 1% for continuous data and < 15% for the categorical data, except for folic acid supplementation use (26%). We used multiple imputation to impute missing values for covariates . PM10 and NO2 exposure averages for corresponding periods were moderately correlated (r = 0.57–0.63). Figure S3 in Supplemental Material (http://dx.doi.org/10.1289/ehp.1003316) presents maps of the distribution of PM10 and NO2 concentrations in the study area, demonstrating substantial spatial differences in annual average concentrations (up to 4–8 μg/m3) between urban and suburban areas.Air pollution and fetal growth characteristics.10 and NO2 levels were not consistently associated with second-trimester or neonatal head circumference, but higher levels were associated with smaller fetal head circumference in the third trimester . PM10 levels were not associated with fetal or neonatal length, but NO2 levels were inversely associated with fetal femur length in the second and third trimester (p < 0.01). Exposure to PM10 was associated with increased estimated fetal weight in the second trimester (p < 0.05), but PM10 and NO2 levels were associated with a lower birth weight .10 and NO2 exposure levels (highest vs. lowest quartile) were associated with reductions of 13 g and 46 g in third-trimester weight and birth weight, respectively, for PM10 and reductions of 20 g and 61 g in third-trimester weight and birth weight, respectively, for NO2 (results not shown). When including both PM10 and NO2 in the models, the inverse association for PM10 exposure with first-trimester crown–rump length reached statistical significance, and the associations for PM10 exposure with third-trimester head circumference and for NO2 exposure with femur length in the second and third trimester persisted (results not shown). The unadjusted associations were consistent with the adjusted associations, although somewhat stronger inverse associations and smaller p-values were observed for PM10 exposure with weight in the third trimester and at birth, and for NO2 exposure with head circumference in the second and third trimester, length neonatally, and weight in the third trimester and at birth (results not shown).When comparing the individual associations of maternal air pollution exposure and smoking during pregnancy with weight by trimester, we observed that inverse associations for air pollution exposure were smaller in magnitude than associations for maternal smoking but still considerable: Smoking compared with no smoking was associated with decreases of 39 g and 146 g in third-trimester weight and birth weight, respectively, whereas elevated PM10 and NO2 exposure with longitudinally measured fetal growth. Compared with the first quartile, the third and fourth quartiles of PM10 and NO2 exposure showed a significant overall difference in head circumference growth during pregnancy  . < 0.01) . No sign < 0.01) . Figure Air pollution and risks of adverse birth outcomes. Compared with mothers in the lowest quartile of PM10 exposure levels, exposures in the third and fourth quartiles were positively associated with preterm birth  , but no significant association was observed for the fourth quartile . No consistent associations were observed for NO2 exposure with adverse birth outcomes. When including both PM10 and NO2 in the models, associations for PM10 exposure with preterm birth became larger in magnitude, and larger effect estimates with smaller p-values were observed for associations between PM10 exposure and low birth weight (results not shown). Relative to the adjusted models, the unadjusted models showed smaller p-values for the positive associations for PM10 exposure with preterm birth and SGA at birth  and larger effect estimates with smaller p-values for the associations for NO2 exposure with preterm birth, low birth weight, and SGA at birth  (results not shown).ctively] . The thi10 and NO2 is inversely associated with fetal growth in the second and third trimester and with weight at birth. Elevated PM10 exposure levels were also positively associated with preterm birth and SGA.Results from this large population-based prospective cohort study from early pregnancy onward suggest that maternal exposure to PMAir pollution, fetal growth, and birth outcomes. A few animal experiments suggested effects of maternal exposure to air pollution on placental function and fetal growth , and sulfur dioxide (SO2) during different periods with fetal growth parameters. No significant associations were observed for NO2 exposure. The authors reported that the observed associations were heterogeneous regarding the specific exposure periods and outcome measures examined  and biparietal diameter growth during the second trimester and between NO2 exposure and SD scores for both size and growth of second- and third-trimester head circumference, abdominal circumference, biparietal diameter, and estimated fetal weight    with hea.5)   have been linked to outcomes such as preterm birth, low birth weight, and intrauterine growth restriction  PDFClick here for additional data file.


Triplets are often born premature and with a low birth weight. Because the incidence of triplet births is rare, there are relatively few studies describing triplet birth weight characteristics. Earlier studies are often characterized by small sample sizes and lack information on important background variables such as zygosity. The objective of this study is to examine factors associated with birth weight in a large, population-based sample of triplets registered with the Netherlands Twin Register (NTR).In a sample of 1230 triplets from 410 families, the effects of assisted reproductive techniques, zygosity, birth order, gestational age, sex, maternal smoking and alcohol consumption during pregnancy on birth weight were assessed. The resemblance among triplets for birth weight was estimated as a function of zygosity. Birth weight discordance within families was studied by the pair-wise difference between triplets, expressed as a percentage of the birth weight of the heaviest child. We compare data from triplets registered with the NTR with data from population records, which include live births, stillbirths and children that have deceased within days after birth.There was no effect of assisted reproductive techniques on triplet birth weight. At gestational age 24 to 40 weeks triplets gained on average 130 grams per week; boys weighed 110 grams more than girls and triplets of smoking mothers weighted 104 grams less than children of non-smoking mothers. Monozygotic triplets had lower birth weights than di- and trizygotic triplets and birth weight discordance was smaller in monozygotic triplets than in di- and trizygotic triplets. The correlation in birth weight among monozygotic and dizygotic triplets was 0.42 and 0.32, respectively. In nearly two-thirds of the families, the heaviest and the lightest triplet had a birth weight discordance over 15%. The NTR sample is representative for the Dutch triplet population that is still alive 28 days after birth.Birth weight is an important determinant of childhood development. Triplet status, gestational age, sex, zygosity and maternal smoking affect birth weight. The combined effects amount to a difference of 364 grams between monozygotic girl triplets of smoking mothers compared to dizygotic boy triplets of non-smoking mothers of the same gestational age. Birth weight in triplets is also influenced by genetic factors, as indicated by a larger correlation in monozygotic than in di- and trizygotic triplets. The incidence of triplet births is rare. In the Netherlands, up to 1980, a triplet birth occurred once per 10 thousand births. The number of triplet births increased after the introduction of assisted reproductive technologies (ART). In 1990, the number of triplet births had increased up to 6 per 10 thousand. From 2000 onwards, triplet birth rates decline again, mainly because of a change in policy of fertility clinics. The Central Bureau of Statistics of the Netherlands has monitored triplet birth rates at 2 per 10 thousand births since 2005 )Starting in 2009, the Teacher Report Form  is collTriplets who agreed to participate in ANTR research have usually been included in the ANTR data collection. The procedure of data collection of the ANTR is described elsewhere ,32.For the current study, we used the data on BW, smoking and alcohol consumption during pregnancy acquired from questionnaire Q1 and Q2, completed by the mother of the triplets. Data on ART came from Q1, Q2 or Qft. Data on BW and zygosity were available for 455 and 465 trios, respectively, out of the 642 complete trios. For 433 trios information on both was available. For 410 trios, data were available for all variables under study, i.e. BW, zygosity, GA, alcohol consumption and smoking. Information on ART was available for 329 out of these 410 trios.Response consistency was investigated by correlating responses given at subsequent questionnaires. When comparing Q1 and Q2, correlation was 0.93 (N = 521) for BW, 0.94 (N = 182) for GA, 0.95 (N = 182) for smoking, 0.57 (N = 179) for alcohol consumption and 0.98 (N = 100) for ART. Correlation for ART between Q1 and Qft was 0.95 (N = 132), and for ART between Q2 and Qft it was 1.00 (N = 132). Except for alcohol consumption, responses were highly consistent. For alcohol consumption, 14 mothers reported to have consumed alcohol on Q1 but not on Q2, and 7 mothers reported to have consumed alcohol on Q2, but not on Q1. In these cases, the response given at Q1 was used in the analysis.Analyses were performed using the software package Mx, which allows modeling of the dependency that exists between measures of pairs of relatives . We test2, sex, alcohol consumption and smoking. In step 2, the means, variances and covariances of the MZ triplets from the MZ group were tested for equality with the MZ triplets from the DZ trio group. In step 3, the means, variances and covariances of the single remaining triplets in a DZ trio, not part of the MZ pair, and TZ triplets were constrained to be equal. In step 4, birth order effects on mean BW were tested within the zygosity groups. Finally, in step 5 it was tested whether the effects of GA, GA2, sex, maternal smoking and alcohol consumption on mean BW significantly differed from zero.First a full model, in which all effects were estimated, was fitted to the data. Subsequently, nested sub models were tested. In the full model the following parameters were estimated: the grand mean BW as a function of birth order and zygosity, the variance and covariance in BW (as a function of zygosity), and the regression coefficients on BW of GA, GAχ² distribution with the degrees of freedom (df) equaling the difference in df between the two models. A p-value of 0.05 was used to determine statistical significance.Using the raw likelihood method as implemented in Mx, the different models were compared using the log-likelihood ratio test. The difference in -2 times the log-likelihood (-2LL) between two nested models has a 2. GA2 was calculated by squaring the normalized score of the variable GA as described above . Smoking and alcohol consumption during pregnancy were analyzed as dichotomous traits: yes (1) or no (0).In the regression analysis, sex was coded 0 for boys and 1 for girls and GA was coded as actual GA minus 40, ranging from 0 (GA of 40 weeks) to -16 (GA of 24 weeks). A possible flattening of triplet BW at the highest GA's was modeled with GAThe MZ and DZ correlations were used to infer the influence of genetic and/or environmental factors on triplet BW. MZ triplet pairs and trios are genetically identical, whereas DZ and TZ triplet pairs and trios share on average 50% of their genetic material. A MZ correlation that is higher than the DZ correlation implies the influence of genetic factors. A DZ correlation that is higher than half the MZ correlation implies that shared environmental factors influence BW.Three definitions are commonly used for BW discordance in twins. First, an absolute definition: the absolute difference in BW. Second, a percentage definition: BW disparity is calculated as a percentage of the largest child's BW. Third, a statistical definition: BW as percentile of one or two standard deviations from the mean . StudiesThe third method, specific to trios, takes into account the BW of the triplet that falls in between the heaviest and lightest triplets. In this method, a relative BW ratio is calculated by taking the difference between the middle and lightest triplet as a percentage of the difference between the heaviest and lightest triplet . An advaWe used the percentage definition to estimate birth weight discordance, as this definition is most frequently used.Table t(454) = 1.90, p = 0.06). There was a difference in maternal education (χ2(3) = 8.69, p = 0.03). Maternal education in the responders group was lower than the educational level of mothers from the non-responders group . There was also a difference in population density between the two groups (χ2(3) = 8.16, p = 0.00). Population density was categorized as more than or less than 1000 persons per square meters. The distribution in the response group was 50.0% and 50.0% and in the non-response group 43.4% and 56.6%, for < 1000 and > 1000 persons per square meters, respectively.Comparing the current study sample with trios registered in the NTR but not participating in this survey yielded no significant age difference between the two groups , the age of mothers who spontaneously conceived the trio ranged from 20 to 41 years . ART was overrepresented in the TZ triplet group. 87% of the mothers of TZ triplets reported ART, compared to 19% of the mothers of DZ's and only 3% of the mothers of MZ triplets.2(1) χ= 0.23, p = 0.63). ART was also tested in the TZ triplet group alone to correct for a possible confounding effect of zygosity, as TZ triplets are possibly heavier and overrepresented in the group of triplets born after ART. Still, no effect was found (2(1) χ= 1.23, p = 0.27). All other analyses were therefore performed on the total set of triplets, including triplets without information on ART.Within the subset of triplets with information on ART, we did not find a significant effect of ART on BW  [th percentile of BW is 1673 grams, but for triplets it is only 1418 grams. The discrepancy between singletons and triplets increases with increasing GA. As such data are not available for Dutch triplets, we present the percentages of triplets who are SGA based on singleton standards for the United States (US) and based on US triplet standards as reported by Alexander in addition to Swedish singleton standards which are comparable to Dutch singleton standards [Table . (1998)  reportedtandards ,40.2(3) χ= 16.57, p = 0.00). A significant birth order effect was found within the TZ group (2(2) χ= 29.07, p = 0.00), but not in the MZ and DZ group (2(2) χ= 5.61, p = 0.06 and 2(2) χ= 0.92, p = 0.63, respectively).Fit statistics of all tested models are presented in Table 2) was observed (2(1) χ= 2.49, p = 0.11). An effect of sex was found with boys being 110 grams heavier than girls (2(1) χ= 36.69, p = 0.00). Furthermore, triplets from mothers who smoked during pregnancy were 104 grams lighter than the triplets of mothers who did not smoke (2(1) χ= 10.9, p = 0.00). We did not find a significant effect of alcohol consumption during the pregnancy on triplet BW (2(1) χ= 0.11, p = 0.74).The tests of the fixed effects showed that GA was the most important contributor to mean BW in triplets. Between a GA of 24 to 40 weeks, the triplets gained 130 grams per week. No significant flattening of the growth line  twin, has an heavier placenta and a more optimal  cord insertion . PossiblThe prevalence of BW discordance is comparable with a study of Jacobs et al. (2003)  and otheWe also estimated the resemblance in BW of triplets. The resemblance for BW in MZ triplets was higher than in DZ triplets. Both were higher before correction for GA, sex and smoking than after adjusting for these factors. This suggests that genetic as well as common environmental factors influence birth weight and that GA, sex and smoking are some of the specific common environmental factors.We observed that, when taking US singleton standards as a reference, 40-50% of the triplets were SGA. Children who are born SGA are at risk for asphyxia and intrauterine mortality ,47. As aPopulation based triplet zygosity distributions have changed over the years. Imaizumi  reportedStudies in singletons report that children born after ART are lower in BW than spontaneously conceived children. In twins the effect of ART is less clear, some studies find an effect, while others do not ,50. The The present sample consisted of triplets who were registered at the NTR and whose parents were willing to participate in survey studies. This led to a small positive selection bias. Triplets from families in which all three children are alive 28 days after birth also have more favorable scores on BW and GA. Moreover, parents are possibly more willing to participate in research when the triplets are healthy compared to parents dealing with illnesses of one or more of their children. The NTR sample was more comparable with a selection of Dutch triplets that were still alive 28 after birth, than with a complete group of Dutch triplets including children who died soon after birth. A comparable positive selection bias was found in a study on secular trends in gestational age and birth weights in twins. In this study twins registered at the NTR were compared with a national reference set. The twins registered at the NTR were found to have a higher GA (36.5 (2.4) compared to 35.9 (3.0) weeks) and a higher BW (2498 (550) grams compared to 2459 (615) grams). As a reWe investigated whether zygosity influenced triplet BW. MZ triplets are more in competition for nutrients than DZ and TZ triplets. A more direct measure of triplets sharing placenta's and therefore triplet nutrients competition is chorionicity. Information on chorionicity would therefore have been a valuable addition to the information on zygosity. Currently, no reliable information on chorionicity was available.Parity has been associated with BW but was not included in the analysis of present study, as information on parity was only available for about three-quarter of the mothers. Including parity would therefore have decreased sample size considerably. In an analysis within the reduced sample, there was no significant effect parity on BW.Some strengths of this study are also noteworthy. The sample is relatively large. We do not know of another study that took so many risk factors into account analyzing their effect on triplet BW. This study is the first to describe triplet zygosity in the Netherlands based on individual zygosity measures instead of population based estimated zygosity distribution. We therefore could confirm the assumption that ART has inflated the Dutch TZ triplet population. Furthermore, this study is the first to investigate the effect of maternal smoking and alcohol consumption during pregnancy on triplet BW. In addition, although our sample is somewhat positively biased when comparing it to all triplets born in the Netherlands, it is a representative sample for the Dutch triplet population that is still alive one month after birth.Longitudinal data collection on triplets is scarce. Data collection within the Netherlands Twin Register (NTR) is broad, including an important focus on behavior. The data collection in triplets that we are currently establishing is unique in its kind. With this dataset it is possible to study long term effects of low BW in triplets, both on physiologic and also on behavioral level. This study was limited to a description of the sex and zygosity distribution of the triplets and the effect of a number of BW characteristics. We found an effect of GA, sex, birth order, zygosity and maternal smoking on triplet birth weight, but found no effect of ART and maternal alcohol consumption. The combined effects implied that differences of 364 grams can be observed between MZ girl triplets of smoking mothers compared to TZ boy triplets of non-smoking mothers of the same GA. Furthermore, we found that MZ triplets resembled each other more than DZ triplets, indicating that, in addition to environmental factors, genetic factors contribute to triplet BW.ART: assisted reproductive techniques; BW: birth weight; DZ: dizygotic; GA: gestational age; MZ: monozygotic; NICU: neonatal intensive care units; NPR: Netherlands Perinatal Registry; NTR: Netherlands Twin Register; TZ: trizygotic.The authors declare that they have no competing interests.DJL: performed the statistical analyses and participated in the design of the study, data acquisition, interpretation of the statistical analyses, and the draft of the manuscript. CMM: participated in the design of the study, interpretation of the statistical analyses, and draft of the manuscript. CEMB: participated in the data acquisition. JMV: participated in the data acquisition. MCH: participated in the interpretation of the statistical analyses. DIB: participated in the design of the study, interpretation of the statistical analyses, and draft of the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2431/11/24/prepub


Many previous studies have found seasonal patterns in birth outcomes, but with little agreement about which season poses the highest risk. Some of the heterogeneity between studies may be explained by a previously unknown bias. The bias occurs in retrospective cohorts which include all births occurring within a fixed start and end date, which means shorter pregnancies are missed at the start of the study, and longer pregnancies are missed at the end. Our objective was to show the potential size of this bias and how to avoid it.To demonstrate the bias we simulated a retrospective birth cohort with no seasonal pattern in gestation and used a range of cohort end dates. As a real example, we used a cohort of 114,063 singleton births in Brisbane between 1 July 2005 and 30 June 2009 and examined the bias when estimating changes in gestation length associated with season (using month of conception) and a seasonal exposure (temperature). We used survival analyses with temperature as a time-dependent variable.We found strong artificial seasonal patterns in gestation length by month of conception, which depended on the end date of the study. The bias was avoided when the day and month of the start date was just before the day and month of the end date (regardless of year), so that the longer gestations at the start of the study were balanced by the shorter gestations at the end. After removing the fixed cohort bias there was a noticeable change in the effect of temperature on gestation length. The adjusted hazard ratios were flatter at the extremes of temperature but steeper between 15 and 25°C.Studies using retrospective birth cohorts should account for the fixed cohort bias by removing selected births to get unbiased estimates of seasonal health effects. Worldwide, it is estimated that 2.2% of all babies are stillborn  and 9.6%Environmental and meteorological factors may be a cause of adverse birth outcomes . IncreasStudies in developing countries have hypothesised that seasonal patterns in birth outcomes are due to seasonal changes in nutrition, infectious diseases, and seasonal work . These fFor birth cohorts the population at-risk is constantly changing, as new pregnancies start and existing pregnancies end . In retrfixed cohort bias.The pattern shown in Figure both the start and end times are unknown. These subjects will be missing from the at-risk population which will therefore be too small, and hence the consequences of the bias are similar to those caused by censoring.The arrows shown in Figure The purpose of this paper is to demonstrate the potential effects of ignoring the fixed cohort bias, and to show how it can be avoided. We show the bias when estimating the effects of season, and the seasonal exposure of temperature.We requested data on a cohort of all singleton births in the Brisbane Statistical District from 1 July 2005 to 30 June 2009  from the Queensland Health Perinatal Data Collection Unit. We only examined singleton births, so twins or triplets were excluded.To investigate the fixed cohort bias we simulated a retrospective birth cohort with no seasonal pattern in gestation. The simulated cohort was based on the Brisbane cohort and had the same start and end dates (1 July 2005 to 30 June 2009). We randomly sampled gestation lengths from the Brisbane cohort (with replacement) using the original sample size . We randomly simulated conception times from 43 weeks prior to the start of the study period (3 September 2004) until the end of the study period using a discrete Uniform distribution. Using these non-seasonal conception times and randomly sampled gestation lengths meant there was no seasonal pattern in gestation lengths or dates of births in the simulated cohorts, and that the distribution of gestation lengths was the same as the Brisbane cohort. Simulated births with a date of birth outside the cohort start and end dates were excluded, as these births would have been missed by a fixed cohort. Missing these births is the cause of the fixed cohort bias. The simulations were made using the R software (version 2.11.1).We ran a survival analysis of gestation length to analyse the effect of the fixed cohort bias on the seasonality of birth outcomes. We used a Cox proportional hazards model with a dependent variable of gestation length (in days) and an independent variable of conception month. A hazard ratio greater than one means an increased chance of giving birth, and hence shorter gestations. Whereas a hazard ratio less than one means a reduced chance of giving birth, and hence longer gestations.We suspected that the effects of the fixed cohort bias would vary depending on the start and end dates of the study. To investigate this, we moved the end date of the study (using the simulated cohort) backwards in time (one month at a time) and repeated the survival analysis. We repeated this analysis for 100 simulated cohorts and calculated the average hazard ratios in each conception month.We propose a method to avoid the fixed cohort bias by removing some pregnancies from the cohort. The adjusted cohort is created by limiting the included pregnancies to those with conception dates between:1. 19 weeks before the cohort started, and2. 43 weeks before the cohort ended.We used 19 and 43 weeks as cut-offs since these were the longest and shortest gestation lengths observed in the Brisbane cohort. This ensures that the long pregnancies at the beginning of the cohort, and the short pregnancies at the end of the cohort, are excluded. We then repeated the survival analysis using the original Brisbane cohort and the adjusted cohort with month of conception as the single independent variable.To investigate the effect of the fixed cohort bias on the effect of seasonal environmental exposures, we fitted a similar survival model but with the time-dependent variable of mean temperature in the last four weeks. We used polynomial splines for mean temperature . The degTable Our simulated data had no seasonal pattern in gestation length. Figure We further illustrate the fixed cohort bias using the Brisbane cohort in Figure The effect of the fixed cohort bias in a study investigating temperature is shown in Figure Artificial seasonal patterns can occur in retrospective birth cohorts depending on the study's start and end dates Figure . The cauThe biases are not caused by the well-known issues of censoring in survival data, which occurs when the start or end time of a subject are unknown. The bias demonstrated here is because both the start and end times are unknown. It is a bias caused by subjects who were completely (and unknowingly) missed from the cohort, not because of subjects who were partially observed.The potential biases due to changes in the at-risk population when examining the effect of time-dependent exposures on pregnancy have been addressed by two recent studies ,22. ThesOur results show how the size of the fixed cohort bias can be substantial, causing great changes in the months that most effect gestation length Figure , 4, and As shown in Figure A group of conceptions that were missed in this study were those occurring before 20 weeks (spontaneous abortions). These unfortunate cases are not added to the Brisbane birth registry. Rates of spontaneous abortions having reached at least five gestational weeks after last menstrual period vary from 11 to 16% -28. MissAn ideal study design is one that prospectively follows women from close to conception, as this ensures that no pregnancies are missed (and hence the fixed cohort bias is not an issue). However, these designs are much more expensive that those relying on routinely collected birth data. A good alternative is to adjust cohorts using our method, and so avoid artificially created patterns and wrongly estimated effects from seasonal environmental exposures on birth outcomes.Our results in Figure Although we controlled for Indigenous status we did not have data on other racial groups. As shown by Darrow et al  seasonalWe used a Cox proportional hazard model with a time-dependent exposure and therefore the results are presented as hazard ratios, and we were not able to give the estimated gestation length on the absolute scale of time Figure . Using aCorrectly estimating the effects of environmental exposures on pregnancy is vital because an unhealthy start to life can mean an unhealthy adulthood. The adverse health effects of preterm birth and low birth weight include: socio-emotional and educational problems , reducedThe authors declare that they have no competing interests.LS and AGB discovered the bias and designed the method to avoid it. LS wrote the first draft of the manuscript. AGB ran the statistical analyses. ST supervised the study and contributed to the final draft. All authors contributed to study concept and design. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2288/11/49/prepub


There is agreement that symptomatic sacral meningeal cysts with a check-valve mechanism and/or large cysts representing space-occupying lesions should be treated surgically. This study investigated factors indicating a need for surgical intervention and surgical techniques for sacral meningeal cysts with a check-valve mechanism.In ten patients presenting with sciatica and neurological deficits, myelography, computed tomography (CT) myelography, and magnetic resonance imaging (MR imaging) detected sacral meningeal cysts with a check-valve mechanism. One patient had two primary cysts. Ten cysts were type 2 and one cyst was type 1. Nine of the ten patients had not undergone previous surgery, while the remaining case involved recurrent cyst. For the seven patients with normal  type 2 cysts and no previous surgery (eight cysts), suture after collapse of the cyst wall was performed. For the recurrent type 2 cyst, duraplasty and suture with collapse of the cyst wall were performed to eliminate the check-valve mechanism. For the remaining type 2 cyst, a primary root was sacrificed because of the huge size of the cyst. For the type 1 cyst, the neck of the cyst was ligated.In all cases, chief complaints disappeared immediately postoperatively and no deterioration of clinical symptoms has been seen after a mean follow-up of 27 months.The presence or absence of a check-valve mechanism is very important in determining the need for surgical intervention for sacral meningeal cysts. A sacral meningeal cyst is often found incidentally, and no specific surgical method has been established for most cases , 14. WitParticipants in this study comprised six men and four women . Myelography, computed tomography (CT) myelography, and magnetic resonance imaging (MRI) were performed preoperatively in all the cases. CT myelography is essential to clarify the bone structure and the relationships with the cyst and the bone. Multiple cysts were observed in all cases, and myelography and CT myelography were required to identify the primary cyst. In nine cases, only one primary cyst was present  were type 2 and one case was a type 1 sacral meningeal cyst. Of the ten patients, nine had not undergone any previous surgery for cysts. The remaining patient showed recurrent cyst. In this recurrent case, cyst-subarachnoid shunt had been performed 8 years earlier, but the shunt system had started to malfunction Fig. . All tenFor the seven patients with normal  type 2 cyst and no previous surgery (eight cysts), suturing the cyst wall after collapse of the cyst wall was performed. For the recurrent case, duraplasty and suture after collapse of the cyst wall were performed to eliminate the check-valve mechanism of the cyst. For the remaining type 2 cyst without previous surgery, the primary root was sacrificed under meticulous monitoring because of the huge size of the cyst Figs.  and 6. FPerioperative complications were not observed in any except the recurrent case. In that case, subcutaneous leakage of cerebrospinal fluid (CSF) occurred 2 weeks after surgical intervention. After spinal drainage was performed for 4 weeks, CSF leakage disappeared completely. In all cases, the chief complaints disappeared immediately postoperatively. All cases were followed-up after the operation , with no further recurrences identified in any cases. In the seven cases (eight cysts) of normal type 2 cyst with no previous surgeries, the check-valve mechanism was eliminated naturally (automatic elimination) after collapse of the cyst wall and suturing. Elimination of the check-valve mechanism could be confirmed when CSF began to flow naturally through the cyst.Due to advances in MRI, sacral meningeal cysts are increasingly beingincidentally , 13, 15.Previous reports have mentioned the use of various operative methods for cysts with check-valve mechanisms , 7, 12. Generally, the check-valve mechanism is eliminated naturally after suturing and collapse of the cyst wall. However, duraplasty must be performed in recurrent cases, when the check-valve mechanism has not been eliminated and surgical removal is required to establish communication with the caudal sac. The presence or absence of a check-valve mechanism is very important in determining whether surgical intervention for the sacral meningeal cyst is necessary.


The Australian baby bonus maternity payment introduced in 2004 has been reported to have successfully increased fertility rates in Australia. We aimed to investigate the influence of the baby bonus on maternal demographics and birth characteristics in Western Australia (WA).This study included 200,659 birth admissions from WA during 2001–2008, identified from administrative birth and hospital data-systems held by the WA Department of Health. We estimated average quarterly birth rates after the baby bonus introduction and compared them with expected rates had the policy not occurred. Rate and percentage differences  were estimated separately by maternal demographics and birth characteristics. WA birth rates increased by 12.8% following the baby bonus implementation with the greatest increase being in mothers aged 20–24 years , mothers having their third  or fourth child , mothers living in outer regional and remote areas , mothers giving birth as public patients , and mothers giving birth in public hospitals . Interestingly, births to private patients  and births in private hospitals  decreased following the policy implementation.The introduction of the baby bonus maternity payment may have served as an incentive for women in their early twenties and mothers having their third or fourth child and may have contributed to the ongoing pressure and staff shortages in Australian public hospitals, particularly those in outer regional and remote areas. We also investigated where the baby bonus had the greatest birth rate influence with regard to other maternal demographics and birth characteristics using data with longer follow-up than previously published research The use of de-identified, administrative health data for this study without patient consent was approved by the Human Research Ethics Committee of the WA Department of Health. This study was performed in accordance with the Declaration of Helsinki.WA birth data from the WA Midwives Notification System (MNS) from July 2001 to December 2008 was linked with the WA Hospital Morbidity Data Collection (HMDC) by the Data Linkage Branch at the WA Department of Health and provided to the researchers in de-identified form. The collection of this data is governed by legislation requiring all births in WA occurring on or after 20 weeks gestation or infants born with birth weight of at least 400 g to be registered as well as all hospital admissions and separations from all hospitals in the State.For the study population included in this study, multiple births (e.g. twins) were counted as one birth admission, with the information on length of hospital stay for the first born twin being used. Also, both live- and stillborn infants were included. Length of stay was categorized into 0–3 days and 4+ days following birth since most mothers and babies stay less than 4 days in hospital following an uncomplicated vaginal birth.The MNS included information on the Index of Relative Socio-Economic (SE) Advantage and Disadvantage (IRSAD) and Accessibility/Remoteness Index of Australia (ARIA+). The IRSAD values are based on information on household income, educational attainment and occupation from the Australian Census conducted every five years and assigned to each collection district area in the state. The IRSAD values were divided into sextiles for all analyses with the highest three sextiles combined to indicate high area-based SES and the lowest three combined to indicate low SES. The ARIA residential remoteness index is also calculated from Census information every five years and reflects access to services in a collection district area. It was divided into major cities, inner regional Australia, outer regional Australia, remote Australia, and very remote Australia. The IRSAD and ARIA values from the 2001 and 2006 Censuses were assigned to each birth admission based on maternal area of residence at the time of birth.The HMDC provided information on the funding source of the mother at the time of each hospital birth. Funding source was categorized to reflect two types of mothers; those treated as public patients and those treated as private patients during delivery. Private patients were defined as those funded with private health insurance (PHI) or who were self-funded, whereas public patients included those insured under the Australian Health Care Agreements and Reciprocal Health Care Agreements. Hospitals were divided into private or public with the exception of the single obstetric tertiary hospital in WA, which was defined as tertiary.We used interrupted time-series analyses to estimate the average quarterly birth rates in WA before (pre-BB) and after (post-BB) the baby bonus implementation and compared the post-BB rates with rates that would have been expected had the policy not occurred (post-2004). We calculated the birth rates shown in nd quarter of 2004 (April 2004-March 2005), during which the baby bonus was announced. This period was excluded from the time series analysis to account for the duration of pregnancy.We assumed the outcome rates followed negative binomial distribution to account for over-dispersion in the data and used segmented regression analyses to measure the impact of the baby bonus We estimated the pre-BB and post-BB average quarterly rates using the segmented regression models and compared the post-BB rates with the expected rates post-2004, calculated from the model as the projection of pre-BB trends under the assumption that no intervention occurred We included 200,659 birth admissions in this study that occurred from July 2001 to December 2008 in WA. The average quarterly birth rates for the pre-BB and post-BB periods and the differences between the post-BB rates and the rates that would have been expected at the same time had the policy not occurred post-2004) are presented in 004 are pIn this study we investigated the influence of the Australian baby bonus introduction on birth rates in WA. Our results indicated that the baby bonus may have served as an incentive particularly for women aged 20–24, women having their third or fourth child and women living in outer regional and remote areas. We found that the increase in birth rates following the baby bonus was greatest in mothers who gave birth in public hospitals or as public patients, for mothers who gave birth vaginally, but with assistance, and for infants who stayed less than four days in hospital following birth.st July 2004 than any other date in the past 30 years in Australia with over a thousand births ‘moved’ so the parents would be eligible for the payment This study used population-based, routinely collected administrative hospital and birth data from WA. We were able to study almost the complete birth information in WA for the time period under study since we received de-identified data from the WA Department of Health for 99.99% of all recorded births for the entire state of WA. Despite obvious strengths relating to the population-based design, we cannot be certain that our findings are due to the baby bonus implementation alone. Some have suggested that the increase in birth rates following the baby bonus introduction could have been explained at least in part by partnering behaviour, economic conditions, unemployment rates, women’s educational attainment, labour force participation, and birth cohort effects Three scientific articles have been published reporting on the effects of the baby bonus introduction on birth rates in Australia Our findings indicate that the baby bonus may have served as an incentive particularly for women aged 20–24, women having their third or fourth child and women living in outer regional and remote areas of Australia. Given that younger women are less likely to hold PHI than older women Public hospitals in Australia have been under pressure during the last couple of decades with medical workforce shortages including shortages of midwives and obstetricians Our results indicate that following the introduction of the baby bonus maternity payment in Australia in 2004, birth rates increased particularly in mothers aged 20–24, mothers having their third or fourth child and mothers living in outer regional and remote areas. The birth rate also increased for mothers giving birth in public hospitals. Considering the reported pressure on Australian public hospitals and ongoing staff shortages, including difficulty in recruiting and retaining obstetricians in rural areas


The objective of the study is to assess maternal factors contributing to under-five mortality at birth order 1 to 5 in India. Data for this study was derived from the children’s record of the 2007 India National Family Health Survey, which is a nationally representative cross-sectional household survey. Data is segregated according to birth order 1 to 5 to assess mother’s occupation, Mother’s education, child’s gender, Mother’s age, place of residence, wealth index, mother’s anaemia level, prenatal care, assistance at delivery , antenatal care, place of delivery and other maternal factors contributing to under-five mortality. Out of total 51555 births, analysis is restricted to 16567 children of first birth order, 14409 of second birth order, 8318 of third birth order, 5021 of fourth birth order and 3034 of fifth birth order covering 92% of the total births taken place 0–59 months prior to survey. Mother’s average age in years for birth orders 1 to 5 are 23.7, 25.8, 27.4, 29 and 31 years, respectively. Most mothers whose children died are Hindu, with no formal education, severely anaemic and working in the agricultural sector. In multivariate logistic models, maternal education, wealth index and breastfeeding are protective factors across all birth orders. In birth order model 1 and 2, mother’s occupation is a significant risk factor. In birth order models 2 to 5, previous birth interval of lesser than 24 months is a risk factor. Child’s gender is a risk factor in birth order 1 and 5. Information regarding complications in pregnancy and prenatal care act as protective factors in birth order 1, place of delivery and immunization in birth order 2, and child size at birth in birth order 4. Prediction models demonstrate high discrimination that indicates that our models fit the data. The study has policy implications such as enhancing the Information, Education and Communication network for mothers, especially at higher birth orders, in order to reduce under-five mortality. The study emphasises the need of developing interventions to address the issues of anaemia, mothers working in the agricultural sector and improving relevant literacy among mothers. Globally, the under-five mortality rates have declined from 85 per 1000 to 51 per 1000 UNICEF . Howevera While birth order plays the role of a strong confounder in under-five mortality, Srinivasan postulated that the intrinsic growth rate and the mean generational length of any population may get affected by the birth order pattern (Srinivasan The United Nations Children’s Fund (UNICEF) identifies that children are at a greater risk of dying before age five if they are born in rural areas, among the poor, or to a mother deprived of basic education UNICEF . Even thData file IAKR52FL.SAV on children records was accessed from Demographic and Health Survey (DHS) site with due permission. The sample was weighted for population differences, rural–urban and slum and non-slum proportional differences to arrive at a nationally representative data in NFHS- 3 conducted on the DHS format used across 80 countries NFHS-3 . The natMother’s age at index child in years, gender of child at index birth order , place of residence , religion , type of caste , standard of living , mother’s occupation , mother’s anaemia level , prenatal care , assistance at delivery , antenatal care , information regarding complications in pregnancy , pregnancy health nutrition education , Breastfeeding health nutrition education , place of delivery , breastfeeding , immunization , child’s age in months , birth interval  and region  are all taken as explanatory variables as defined by the NFHS-3. For purposes of univariate and multivariate analysis, a few variables are modified/ merged into lesser categories for better understanding of results. Wealth index  is recoded as poorest and poor, middle and richer and richest. Mother’s and father’s education is coded as no formal education and primary, secondary and greater than higher secondary. Father’s occupation  is taken as part of the definition of the socio-economic status of the family. Size of child at birth  are recoded as larger than average, average and smaller than average. All these variables are taken as independent variables and child’s status (live or dead) as dependent variable in the study. Child’s age is calculated as difference in months between the date of birth and the date of interview for living child and between the date of birth and the date of death of child as reported by the mother NFHS-3 .Frequency tabulations describe the characteristics of under-five children at each birth order in the study. It is observed that most of the deaths in the data occur within one month or at the time of birth. Cox Proportional Hazard assumptions assume a linear relationship exists between the endpoint and predictor variables. Further, recall bias would be present in the data due to the retrospective nature of the data collection. Due to non-fulfillment of assumptions and the possibility of recall bias in the current data set that could cause misleading estimates, logistic regression is chosen as the appropriate statistical tool for analysis. Logistic regression is used to determine factors associated with the mortality forcing child age in months and mother’s age at index child into multivariate analysis at each birth order as known confounders. Significant variables at p-value ≤ 0.10 and modified appropriate categories are considered for multivariate logistic regression analysis after checking co-linearity among the explanatory variables. Forward stepwise logistic regression methods are applied to assess significant explanatory variables of under-five mortality for each birth order. Cross-validation of confidence intervals calculated for odds ratios derived by multivariate logistic regression analysis are internally validated via bootstrapping re-sampling methods using 100re-samples for each birth order model  is taken as dependent variable and percentage of deaths in parenthesis with distribution of each category of variable for each birth order are provided in Table Unadjusted odds ratios and 95% CI of Univariate analysis are presented in Table In Birth Order 2, mother’s education and occupation, mother’s anaemia, place of residence, caste, breastfeeding, immunization, wealth index, father’s education, standard of living, place of delivery, size of child at birth, prenatal care and previous birth interval greater than 2 years are significant factors. In Birth Order 3, mother’s education, religion, mother’s anaemia, place of residence, caste, breastfeeding, wealth index, father’s education, standard of living, place of delivery, size of child at birth and birth interval are significant factors.In Birth order 4, mother’s education, religion, gender, breastfeeding, wealth index, standard of living, size of child at birth and birth interval are statistically significant. Breastfeeding, wealth index and standard of living increase the chance of under-five survival. A below average size of child at birth increases the risk of dying by two and a half times compared to an above average size of child at birth . A birth interval of greater than 24 months increases under-five survival significantly. Muslim religion acts as a protective factor in under-five survival compared to Hindu religion. Female children are at a 26%  greater risk of dying than male children.In Birth Order 5, religion, mother’s anaemia, gender, breastfeeding, size of child at birth and birth interval are statistically significant. The most notable absences at birth order 5 are mother’s and father’s education. Muslims have a 33%  better chance of survival than Hindus. Others have a risk factor at 14%  compared to Hindus but is not statistically significant. Mother’s anaemia effects under-five survival inversely. A female child is at 48%  greater risk of mortality than a male child. Breastfeeding increases the chance of under-five survival. A below average size of child at birth increases the risk by 2 times  compared with above average size of child at birth. A birth interval greater than 24 months increases under-five survival compared to lesser birth interval.Most fathers whose children died are poor, with low standard of living, no formal education and working in agricultural sector for birth order 1 and 2. However, fathers either working or in agriculture had similar effects for birth order 3 to 5. There are very few fathers who are non-working and this did not show any effect on under-five mortality.Adjusted odds ratios with 95% CI for multivariate analysis are provided in Table Bootstrapping analysis showed negligible bias in estimated parameters. The ROC curves for different models are shown in Figure Increasing maternal age is a protective factor in under-five survival. Average maternal age is significantly high for under-five survival in birth order 1 to 5. This is a good sign in a country that has been fraught with issues of child marriages and early and/ or unwanted pregnancies. Almost half of 20–24 year old women in India (44.5%) are married before age 18, and 22% of all 20–24 year old women have given birth by age 18 years NFHS-3 . HoweverThe effects of breastfeeding and birth interval greater than 24 months are significant protective factors from birth order 2 to 5. This is in line with the general perception and is a well- documented fact  compared with scheduled castes. However, it is not a statistically significant factor in multivariate analysis for any birth order. Religion acted differentially in different birth orders. Christians have a better chance at survival in birth order 1 and 3 while Muslims survive better at birth order 4 and 5, compared with Hindus. Multivariate analysis supported univariate analysis with Christians and Muslims having better survival than Hindus in birth order 1 and 4, respectively. This agrees with other studies who found that Hindus were disadvantaged in securing the survival of their babies at early ages compared to Muslims and women belonging to other religious groups  to review the determinants of under-five mortality with a view to improve maternal health care services in India. The predictive accuracy of developed models was above 80% for all developed models indicating a high accuracy to predict the outcome.2 of 0.56, 0.37, 0.33, 0.34 and 0.34 for the developed models 1 to 5, respectively, indicating small effect.The data is based upon recall by the mother and is open to recall biases and undocumented social pressures that lead them to say the “right” thing. The effects of exclusive breastfeeding and total versus partial immunization could not be commented upon due to the restrictions of including breastfeeding and immunization as yes and no dichotomous variables in the analysis. Nutritional differences could also be responsible for differences in under-five survival in religious groups, Christians and Muslims faring better than Hindus. However, our study did not establish a direct link between nutrition and religion though it noted that pregnancy health nutrition education acted as a protective factor. Although contribution of the explanatory variables are statistically significant, predicting mortality accounted for a Nagelkereke RThe study reveals the determinants of under-five mortality according to birth order. While some factors such as breastfeeding and birth interval greater than 24 months are important in under-five survival, several others such as mother’s education and rural- urban place of residence show weaker relationship at higher birth orders. Efforts are needed to educate young women in matters of reproductive health and the benefits of delaying pregnancies. Maternal education, both in formal and informal sectors, can improve their utilization of antenatal care services. Information regarding complications in pregnancy acts as a protective factor for this reason. Pregnancy Health Nutrition education improves under-five survival for birth orders 2 to 5. It is felt that prenatal care and general anxiousness related to first pregnancy that leads couples to seek early care is important in under-five survival. There is need to educate couples regarding correct nutrition, benefits of breastfeeding and immunization and possible complications so that the couples can decide and take precautions. However, complacency at higher birth orders leads to higher rates of under-five mortality. Higher Wealth index is an important indicator in under-five survival. Most parents are not specific about child’s gender in the first birth order but more girl children die at higher birth orders. This clearly points to how gender preferences play a role in under-five mortality. The study identifies various focus groups for increased counselling such as mothers working in the agricultural sector and severely anaemic mothers. However, the study also shows that under-five mortality is as high in rural as urban areas for higher birth orders. It points to the need of IEC  network to be further strengthened to reduce under-five mortality across the board for higher birth orders.No identifying information is available in the data precluding the need for any ethical approvals. We wish to thank the DHS for making this data available for our study.aBirth order (or parity) refers to the numerical order of the live birth or foetal death, recorded in relation to all previous issue of the mother, irrespective of whether the issue is a live birth or foetal death or whether pregnancies were nuptial or extra-nuptial.


Exposure to air pollutants is suggested to adversely affect fetal growth, but the evidence remains inconsistent in relation to specific outcomes and exposure windows.Using birth records from the two major maternity hospitals in Newcastle upon Tyne in northern England between 1961 and 1992, we constructed a database of all births to mothers resident within the city. Weekly black smoke exposure levels from routine data recorded at 20 air pollution monitoring stations were obtained and individual exposures were estimated via a two-stage modeling strategy, incorporating temporally and spatially varying covariates. Regression analyses, including 88,679 births, assessed potential associations between exposure to black smoke and birth weight, gestational age and birth weight standardized for gestational age and sex.st (7.4 μg/m3) to the 25th (17.2 μg/m3), 50th (33.8 μg/m3), 75th (108.3 μg/m3), and 90th (180.8 μg/m3) percentiles, the adjusted estimated decreases in birth weight were 33 g (SE 1.05), 62 g (1.63), 98 g (2.26) and 109 g (2.44) respectively. A significant interaction was observed between socio-economic deprivation and black smoke on both standardized and unstandardized birth weight with increasing effects of black smoke in reducing birth weight seen with increasing socio-economic disadvantage.Significant associations were seen between black smoke and both standardized and unstandardized birth weight, but not for gestational age when adjusted for potential confounders. Not all associations were linear. For an increase in whole pregnancy black smoke exposure, from the 1The findings of this study progress the hypothesis that the association between black smoke and birth weight may be mediated through intrauterine growth restriction. The associations between black smoke and birth weight were of the same order of magnitude as those reported for passive smoking. These findings add to the growing evidence of the harmful effects of air pollution on birth outcomes. Evidence on the potential adverse impact of ambient air pollution on the health of adults and children has grown rapidly over the last two decades. It is now established that short and long-term increases in ambient air pollution are associated with increased mortality and morbidity in adults and children -4. A groThe aim of this study was to investigate the potential associations between individual weekly black smoke exposure estimates and birth weight, gestational age and birth weight standardized for gestational age and sex, in an urban setting. We used data from the United Kingdom (UK) Particulate Matter and Perinatal Events Research (PAMPER) study, set in a single conurbation in northern England (1962-1992) with a largely stable population. The PAMPER study is one of only a few to estimate individual-level exposure to particulate matter during pregnancy using a combination of monitored levels of black smoke (i.e. particulate matter with an aerodynamic diameter of 4 micron or less) and detailed pollution source information.The PAMPER birth cohort consists of all singletons born during 1961-1992 to mothers resident in the city of Newcastle upon Tyne in the north of England . The popAn electronic database of birth records was constructed using data from several sources. The main source was paper-based neonatal records from the two major maternity hospitals in Newcastle: The Princess Mary Maternity Hospital, where neonatal records were available for the entire study period, and Newcastle General Hospital, where records were available from 1967 onwards. Neonatal records from both hospitals contained data on important maternal and fetal/infant characteristics, including maternal name, home address at delivery, baby's name, sex, date and time of birth, gestational age, birth weight and survival status at birth. Data from 'birth ledgers', which are essentially lists of all births whether born in hospital or not, but also containing information on mother's current surname and residential address and the sex, date and place of birth and vital status of the baby, covering 1961-1973 were also included. Additional birth records from Newcastle General Hospital were accessed to complete some key variables which were not in the birth ledgers, including birth weight and gestational age. Home births, except those for the few where the mother was admitted to a maternity hospital shortly after delivery, were only recorded on the birth ledgers and hence had missing birth weight and gestational age information. Birth weights were standardized for gestational age and sex by calculating the z-score of birth weight for gestation separately for males and females using local, northern England, birth weight for gestational age standards . This reAll births were assigned spatial identifiers (zipcodes and/or grid references). For births prior to 1970 (when zipcodes were introduced in the UK), the address at birth was assigned a zipcode according to those in place in 1991 and converted to a grid reference . Where zipcodes could not be completely identified, a grid reference was allocated directly. Grid references allowed linkage with UK census data. Through this, Townsend deprivation scores, an area-based measure of socio-economic status, incorporating home and car ownership, unemployment and overcrowding , were ca4, were obtained from routine data recorded at 20 air pollution monitoring stations within the city boundary between October 1961 and December 1992 and available from the UK Air Quality Archive. The city boundary used was based on the zipcodes NE1-NE7 and part of NE15, excluding Throckley, with the River Tyne forming the Southern boundary. An exact map of the study area is available in a previous publication [Black smoke is a historic measure of airborne particulate matter. Daily black smoke levels, with black smoke approximately equivalent to PMThe modeling process to estimate individual exposure estimates is described in detail elsewhere . Briefly3 are given with corresponding 95% confidence intervals (95% CI). Additionally, if there was evidence of non-linear associations, an alternative functional form was established using fractional polynomials [st, 25th, 50th, 75th and 90th). Interactions between variables were assessed within the linear regression modelling framework. P-values are two-sided. The statistical software package STATA, version 9,  was used for all analyses.Linear regression was used to examine potential associations between black smoke and the three outcome variables , adjusting for year of birth, maternal age, parity, Townsend deprivation score, sex and in the case of birth weight, gestational age. Birth weight and gestational age were treated as continuous variables, maternal age (in five-year groups), parity and sex were treated as categorical. From these models, coefficients per 10 μg/mynomials . For sucThe study received a favorable ethical opinion  from the Sunderland Local Research Ethics Committee (SL REC 1071).A total of 109,086 singleton births to mothers resident in Newcastle for the period 1961-1992 were identified [Table 3 in the 1960s to around 20 μg/m3 in the 1990. Black smoke exposures and year of birth were highly correlated . The seasonal variation of black smoke persisted throughout the study period [3 occurred before 1970 and all below 50 μ/gm3 after 1968.Table y period . The vary period . All indThere were significant inverse linear associations between weekly black smoke exposure and all three outcomes when using the average weekly exposures across the whole of pregnancy, which remained for both birth weight variables following adjustment for other covariates, but not for gestational age Table . Similar1x-2+ β2x-0.5, and for the first, second and third trimesters, the best fitting functional forms were x-1, ln(x) and x-0.5, respectively, where x denotes weekly average black smoke averaged over the entire duration of pregnancy or the corresponding trimester. For an increase in whole pregnancy exposure, from the 1st (7.4 μg/m3) to the 25th (17.2 μg/m3), 50th (33.8 μg/m3), 75th (108.3 μg/m3), and 90th (180.8 μg/m3) percentiles the adjusted estimated decreases in birth weight were 33 g (SE 1.05), 62 g (1.63), 98 g (2.26) and 109 g (2.44) respectively. For an increase in trimester 1 exposure, from the 1st percentile (6.7 μg/m3) to the 25th (16.5 μg/m3), 50th (32.4 μg/m3), 75th (94.1 μg/m3) and 90th (189.5 μg/m3) percentiles, the adjusted estimated decreases in birth weight were 21 g (SE 0.68), 28 g (0.91), 32 g (1.06) and 33 g (1.10) respectively.The associations between black smoke and unstandardized birth weight were significantly non-linear for whole pregnancy and each trimester. For whole pregnancy, the best fitting fractional polynomial model was βst percentile (6.3 μg/m3) to the 25th (16.5 μg/m3), 50th (31.64 μg/m3), 75th (94.8 μg/m3) and 90th (186.7 μg/m3) percentiles, the adjusted estimated decreases in birth weight were 17 g (3.15), 28 g (5.28), 48 g (8.87) and 60 g (11.1) respectively.For an increase in trimester 2 exposure, from the 1st percentile (6.1 μg/m3) to the 25th (15.9 μg/m3), 50th (30.1 μg/m3), 75th (91.0 μg/m3) and 90th (182.4 μg/m3) percentiles, the adjusted estimated decreases in birth weight were 25 g (SE 1.77), 36 g (2.56), 49 g (3.46) and 54 g (3.82) respectively.For an increase in trimester 3 exposure, from the 1st (7.4 μg/m3) to the 25th (17.2 μg/m3), 50th (33.8 μg/m3), 75th (108.3 μg/m3), and 90th (180.8 μg/m3) percentiles the adjusted estimated decreases in standardized birth weight were 0.04 (0.01), 0.07 (0.01), 0.12 (0.02) and 0.14 (0.03) z-score units respectively..For standardized birth weight, the log of exposure β × log(x) was the best fitting fractional polynomial form for the whole pregnancy exposure, while linear models were used for exposures by trimester, because there was no evidence of non-linear associations. For an increase in whole pregnancy exposure, from the 1P = 0.003) and for trimesters 2 and 3 (P ≤0.01. The estimated reduction in birth weight was 17 g (SE ± 9) for the 1st quintile (assumed to be most advantaged), 23 g (SE ± 10) for the 2nd, 42 g (SE ± 9) for the 3 rd, 27 g (SE ± 10) for the 4th quintile and 8 g (SE ± 11) for the 5th quintile (assumed to be the least advantaged) when exposure increased from the 1st percentile (7.4 g/m3) to the 25th percentile (17.2 g/m3) of whole pregnancy black smoke exposure. No other significant interactions were observed. In particular, there was no interaction between black smoke and time period on birth weight when considering the periods before (1962-77) and after (1978-92) the implementation of smokeless zones (P = 0.76).There were significant interactions between black smoke and Townsend deprivation score on both birth weight and standardized birth weight for the whole pregnancy period  increase per 10 μg/m3 increase in exposure.. Significant adjusted associations were seen for gestational age by trimester, corresponding to increases of 0.004 weeks per 10 μg/m3 increase in black smoke for trimester 1, 0.003 weeks for trimester 2 and a decrease of 0.01 weeks for trimester 3  increase per 10 μg/mIn this study, using individual-level estimates of black smoke exposures during pregnancy, associations were seen between black smoke and birth weight, gestational age and birth weight standardized for gestational age. However, not all associations were significant and, regardless of whether significant, effect sizes tended to be fairly small, in particular for gestational age.In contrast to most previous studies which have been ecological in design, with relatively crude measures of exposure, this study utilised individual-level exposure estimates on a large unselected birth cohort. Completeness of the PAMPER database for both the number of births and collected information for each birth is one of the evident strengths . The stuResidential mobility during pregnancy may be associated with exposure misclassification and therefore may introduce bias . Whilst While many studies investigate influences on birth weight, or influences of birth weight on later health, many do not account for the impact of gestational age, a major determinant of birth weight. In this study, in addition to crude birth weight, a standardized measure was also used to give a better measure of fetal growth than birth weight alone. Different methods for gestational age assessment  throughout the study period may introduce bias in gestational age estimation over time. Gestational age in this study was made as objective and accurate as possible by accepting gestational age calculated from the recorded estimated date of delivery  for the majority of births, rather than by using gestational age as recorded in the neonatal notes or birth records. However, these methods are still prone to bias. Most home births in this study period were excluded due to a lack of outcome data. As the majority of home births were in the earlier decades of the study, they would have been associated with relatively high black smoke exposures compared to births in more recent years. However, due to the lack of information from birth ledgers, it was not possible to assess whether this introduced any bias.Adjustment for year of birth in long-term studies of air pollution and birth outcomes is crucial. However, year of birth and black smoke exposure during pregnancy are highly correlated. While it is possible that adjustment for year of birth may reduce the strength of the association between black smoke and birth weight, it is also possible that residual confounding by year of birth or by other factors affecting birth weight and related to year of birth, for example increase in maternal overweight and improvement in quality of pre-natal care over the study period, may remain. There was a striking decrease in gestational age over the study period due to a number of reasons unrelated to air pollution. For example, the higher prevalence of obstetric interventions, including caesarean section, will account for much of the temporal decrease in gestational age.10 concentrations in New Delhi were reported in 2005 to be above 150 μg/m3, and in Beijing above 100 μg/m3 [The composition of black smoke is known to have varied over the study period , reflect00 μg/m3 .10 and birth outcomes: the nature of black smoke measurements means that their conversion factor to PM10 varies both over time and by composition [10 in the context of health studies concluded that daily average black smoke was a reasonable predictor of daily average PM10 [3 increase in average weekly black smoke exposure over the whole pregnancy period was lower than previous estimates , but the estimates were more comparable in the non-linear models.Two considerations are of relevance when comparing estimates of associations between black smoke or PMposition . For theage PM10 . In the 3 with increasing socio-economic deprivation. Industrial processes in the UK cluster in areas of socio-economic deprivation, but attributing such unequal distribution to unequal pollution impact is complicated [There was a significant interaction between community-level socio-economic status and black smoke on birth weight, with increasing estimated decreases in birth weight per 10 μg/mplicated . Other cA pooled analysis showed that passive smoking reduced birth weight on average by 31 g, a meta-analysis reported a 28 g reduction ,34. The Although the association between black smoke and gestational age was not statistically significant overall, it was for each trimester. Given the inconsistent directions between the trimester associations, it is likely that they cancelled out an association for the whole of pregnancy, although the estimated effect sizes were very small. Given the associations between black smoke and both birth weight and standardized birth weight were stronger, it is more likely that the observed associations point towards intrauterine growth restriction as a potential causal pathway for the black smoke effect.Slama et al.  summarisThe findings of this study push forward the hypothesis that the association between black smoke and birth weight may be mediated through intrauterine growth restriction. The estimated effects of black smoke on birth weight were found to be in the same order of magnitude as those for passive smoking exposure. Findings from this study contribute to the growing evidence regarding the adverse effect of ambient air pollution on birth weight. Future reviews of air quality standards should include assessments of fetal growth.10: Particulate matter with an aerodynamic diameter of 10 micron or less; SD: Standard deviation; SE: Standard error; UK: United kingdomCI: Confidence interval; PAMPER: Particulate matter and perinatal events research; PMThe authors declare that they have no competing interests.The PAMPER study was conceived by LP and TPM, who designed this study with MSP, SVG, JR and MC. Outcome data collection was led by SVG and geographical data collection by MC. Analysis was done by RG, under the supervision of MSP and SR. MSP wrote this manuscript with assistance from all other authors. All authors read and approved the final manuscript.