Socioeconomically disadvantaged children are more likely to be of shorter stature and overweight, leading to greater risk of obesity in adulthood. Disentangling the mediatory pathways between socioeconomic disadvantage and childhood size may help in the development of appropriate policies aimed at reducing these health inequalities. We aimed to elucidate the putative mediatory role of birth weight using a representative sample of the Scottish population born 1991–2001 . Estimated height and overweight/obesity at age 4.5 years were related to three measures of socioeconomic disadvantage . Mediation was examined using two approaches: a ‘traditional’ mediation analysis and a counterfactual-based mediation analysis. Both analyses identified a negative effect of each measure of socioeconomic disadvantage on height, mediated to some extent by birth weight, and a positive ‘direct effect’ of mother’s education and Scottish Index of Multiple Deprivation on overweight/obesity, which was partly counterbalanced by a negative ‘indirect effect’. The extent of mediation estimated when adopting the traditional approach was greater than when adopting the counterfactual-based approach because of inappropriate handling of intermediate confounding in the former. Our findings suggest that higher birth weight in more disadvantaged groups is associated with reduced social inequalities in height but also with increased inequalities in overweight/obesity. The existence of social inequalities in health is well established, with poor health disproportionately burdening those of lower socioeconomic status . There iIn this paper we consider childhood growth in terms of age-specific height and overweight/obesity (hereafter ‘overweight’). Height in childhood is often considered as a marker of development, with long-term consequences for future health . In BritThe association between lower birth weight and socioeconomic disadvantage has been extensively documented , 30, 31.The aim of the current study was to elucidate the putative mediatory role of birth weight in the relationship between socioeconomic disadvantage and childhood growth in terms of height and overweight using a large-scale representative sample of the Scottish population. Furthermore, we aimed to provide a comparison of two different methods for examining mediation.The Scottish Longitudinal Study (SLS) is an anonymised record linkage study covering a 5.3% sample of the Scottish population, selected using 20 birth dates . In contThe sample considered for the present analysis are the 43,286 SLS members born from 1991 to 2001. Many analysis variables were from data sources other than the Census: birth registrations, Scottish Morbidity Records, maternity records and Child Health Systems Programme (CHSP) Pre-School data.The SLS contains no identifiable individual level data, and data are derived from linkages that are anonymised prior to analysis by the research team. Ethical approval for the study was granted by the London School of Hygiene and Tropical Medicine Research Ethics Committee (reference 6418) and by the National Health Service National Services Scotland Privacy Advisory Committee for approval of the health data linkage.Three different measures of socioeconomic disadvantage were considered as the exposures of interest: mother’s education; an ecological measure using the Scottish Index of Multiple Deprivation (SIMD); and a synthetic measure of weekly household income. These variables were chosen to cover different domains of socioeconomic disadvantage . Categorical measures of each socioeconomic disadvantage variable were used in the analyses to allow for potential non-linear effects.Mother’s education was derived from 2001 Census data and categorised as: ‘no qualifications’, ‘GCSE or equivalent’ , ‘A-level or equivalent’ , ‘degree or equivalent’.The SIMD defines small area concentrations of multiple deprivation within Scotland across several different domains . SIMD vaIncome data are often missing or poorly measured in observational studies due to their inherent complexity and sensitivity. One solution, recently proposed by Clemens and Dibben , is to dThe CHSP Pre-School is a programme of pre-school child health reviews carried out by nurses and health visitors in Scotland at a series of designated ages. The CHSP Pre-School was established in 1991 but had a phased implementation across the 15 Health Boards in Scotland. Ten Health Boards  had implemented the CHSP Pre-School by 2000  and wereLength/height, weight and age data were extracted from the CHSP Pre-School records at the following approximate ages: 6–8 weeks, 8–9 weeks, 21–24 months, 39–42 months and 48 months. It should be noted that in those Health Boards where the implementation of the CHSP Pre-School was relatively late , some or all of the anthropometric measurements were not available for SLS members born early in the period being considered.The mediator of interest was birth weight . Birth weight was considered as a categorical variable in the analyses to allow for potential non-linear effects.Several background confounding factors were considered: sex, year of birth , Health Board (10 regions) and ethnicity .Intermediate confounders are common causes of the mediator and outcome that are themselves causally affected by the exposure. We considered maternal age at the birth of the study child  and parity prior to the birth of the study child  as intermediate confoundersThe hypothesised causal relationships between the analysis variables are shown in A two-stage modelling approach was used to first define the outcomes and then perform mediation analysis. In the first stage the repeated anthropometric measurements were modelled to predict height and weight at the age of 4.5 years. This age was chosen because it is the approximate mean age at the final CHSP Pre-School measurement. In the second stage, mediation of the effect of the different measures of socioeconomic disadvantage on height and overweight at age 4.5 years by birth weight was assessed using two different approaches: i) ‘traditional’ mediation analysis and ii) counterfactual-based mediation analysis.2). The age- and sex-specific international overweight cut-offs of Cole et al [2 for males and 17.19 kg/m2 for females) were then used to define overweight at age 4.5 years (a binary variable) from the predicted BMI. Predicted height and predicted overweight status were the outcomes of interest to be used in the second stage.Full details of the growth modelling are given in le et al  (17.47 kThis step was achieved using two different approaches: i) ‘traditional’ mediation analysis and ii) counterfactual-based mediation analysisA frequently used approach to mediation analysis, referred to here as a ‘traditional’ mediation analysis and elsewhere as the ‘difference method’ , is thatHowever, this approach is only valid for linear models which do not include exposure-mediator interactions or intermediate confounding –48. We aFor the traditional approach, each derived growth outcome was related to the different indicators of socioeconomic disadvantage using linear or logistic regression (for height and overweight respectively). For each socioeconomic disadvantage measure the following two models were fitted using maximum likelihood estimation:Adjusted for background confounders ;Additionally adjusted for the mediator, birth weight.The regression coefficient for socioeconomic disadvantage in the first fitted model thus gives the estimated ‘total effect’, and the same parameter in the second fitted model gives the estimated ‘direct effect’, with the difference between them giving the estimated ‘indirect effect’. Model specifications were investigated by separately testing the significance of the interaction between the exposure of interest and each covariate (confounders and birth weight) using Wald tests. The ‘proportion mediated’ was then calculated (on the log-odds ratio (OR) scale for overweight models).According to the literature, we would expect both a negative direct effect of socioeconomic disadvantage on height and a negative indirect effect of socioeconomic disadvantage on height, resulting in ‘consistent mediation’ . On the To account for between-subject differences in the precision of the estimated growth outcomes the analyses were weighted by the inverse of the approximate variance of the predicted growth outcome. A robust standard error estimator was used.As noted above, traditional mediation analysis has several limitations. In addition to its restriction to linear models, of particular concern in the present study is that intermediate confounding cannot be appropriately handled. As shown in An alternative approach that can appropriately handle non-linear models and intermediate confounding has been proposed in the causal inference literature. It uses counterfactual definitions of direct and indirect effects allowing for generality and greater formality than the traditional approach . Among sXi be socioeconomic disadvantage, Yi be height at age 4.5 years, Mi be birth weight, Ci be the set of background confounders  and Li be the set of intermediate confounders  for child i. Let Yi(x) be the value that Yi would take if Xi had been set (possibly counter to fact) to the value x, Yi be the value that Yi would take if Xi and Mi had been set to the values x and m, and Mi(x) be the value that Mi would take if Xi had been set to the value x.Let X on Y, conditional on C = c, expressed as a mean difference comparing X = x* to X = x isX on Y, conditional on C = c, expressed as a mean difference comparing X = x* to X = x isX on Y conditional on C = c, expressed as a mean difference comparing X = x* to X = x isThe ‘total causal effect’ (TCE) of The TCE, NDE and NIE can be analogously defined on the OR scale, as shown in Identification of these effects requires certain assumptions. Those usually invoked are non-interference, consistency, and no unmeasured confounding of the exposure-mediator, exposure-outcome and mediator-outcome relationships , 54. AddEstimation was performed via parametric G-computation using Monte Carlo simulation under the additional assumption of correct model specification –56. The Confidence intervals (CIs) were obtained using a bootstrap approach with 1000 draws. The proportion mediated was calculated using the same approach as in the traditional mediation analysis.Analyses were restricted to complete records after assessing the extent of the likely selection bias. All analyses were conducted using Stata version 12 . EstimatGrowth data availability by age and by subject is shown in There was evidence of a strong, graded association between all measures of socioeconomic disadvantage and predicted height at age 4.5 years in the confounder adjusted models with, for example, the most disadvantaged category for mother’s education estimated to be associated with a 0.98 cm  lower height relative to the least disadvantaged category . The estThere was no evidence of interactions between the exposures of interest and either birth weight or the confounders in any models.The TCEs for each measure of socioeconomic disadvantage estimated by G-computation  were verIn this large-scale representative sample of the Scottish population we have elucidated the mediatory role of birth weight in the relationship between socioeconomic disadvantage and childhood growth in terms of height and overweight. Using two different approaches  we found broad agreement that: i) there was a strong, graded, negative ‘effect’ of each measure of socioeconomic disadvantage on height at age 4.5 years which was mediated to some extent by birth weight and ii) there was a strong, graded, positive direct ‘effect’ of each measure of socioeconomic disadvantage (with the exception of synthetic income) on overweight at age 4.5 years which was partly masked by inconsistent mediation by birth weight.These findings suggest that increased birth weight in more disadvantaged groups may be associated with reduced social inequalities in height but also with increased inequalities in overweight. Birth weight could potentially be a modifiable target of intervention through, for example, programmes designed to improve maternal nutrition or deter maternal smoking during pregnancy , 59, 60.Several assumptions must hold in order for us to obtain unbiased estimates of the NDE and NIE using G-computation, in particular: i) no unmeasured exposure-mediator confounding, ii) no unmeasured exposure-outcome confounding, iii) no unmeasured mediator-outcome confounding, iv) correct specification of the parametric models. We believe that we included the main exposure-mediator-outcome confounders  and additional mediator-outcome confounders , which should reasonably capture biological, geographical and temporal sources of confounding, so are hopeful that assumptions i), ii) and iii) hold. All covariates entered the models as categorical variables so there was limited scope for model misspecification and we found no evidence of exposure-mediator or exposure-confounder interactions in any models, so these were not included (assumption iv)).For continuous outcomes, in the absence of intermediate confounding (but with the above assumptions holding) the direct and indirect effects estimated under the traditional mediation analysis are consistent estimates of the NDE and NIE . When coComparison of the results relating to overweight at age 4.5 years between the two analysis approaches is less straightforward. Even if the above assumptions hold and there is no intermediate confounding then the ‘direct effect’ and ‘indirect effect’ estimated under the traditional mediation analysis are not consistent estimates of the NDE and NIE due to the non-collapsibility of the OR . It has Our finding of shorter childhood stature in more disadvantaged socioeconomic groups is in line with many previous studies –10, as iAlthough much previous research has suggested birth weight as a plausible mediator of the effect of social disadvantage on height or overweight, few studies have explicitly examined this mediation. Armstrong et al  used datThe CHSP Pre-School was implemented in different years in different Health Boards , meaning that not all births between 1991 and 2001 were eligible to have all childhood anthropometric measurements taken. Those measurements which could not be observed as they preceded the implementation of the CHSP Pre-School are systematically missing. The growth models required a minimum of one observation so that as many children as possible contributed to the analysis, with information borrowed from other children with more complete growth data. The lack of evidence for an interaction between Health Board and any of the exposures with respect to either of the outcomes suggests that our results remain representative of the target population.Although the analyses considered relatively recent births (1991–2001), the fast-changing nature of the childhood obesity epidemic may mean that the findings are less applicable to current births. Further research is required in more recent cohorts.There are many strengths to this analysis. The large sample size meant that we could estimate effects precisely. Although the analysis sample was a relatively small proportion of the overall sample the main reason for this  should not result in a biased sample–as borne out by the similarity of distributions of variables between the analysis sample and the overall sample. The data were from reliable sources .There were also some limitations to the analysis. Although we included the main confounders and intermediate confounders, there may be others which we were unable to fully capture, potentially leading to bias. Potential unmeasured confounders could include maternal birth weight, parental anthropometrics, maternal lifestyle factors, and maternal morbidity. There may also be residual confounding due to measurement error or lack of granularity in those confounders that we did include.Some of the variables included in the analyses may be subject to measurement error. This is a particular concern for synthetic income, for which error may be due to either the observed data on occupation or to the model from which this is converted into synthetic income. Such measurement error would in general lead to attenuation of the association between synthetic income and height or overweight, which may lead to some bias in the estimated direct and indirect effects.Including a relatively small number of subjects with as few as a single growth measurement, allowing us to retain a larger sample size and reduce the potential for selection bias, may lead to mischaracterisation of the growth trajectories in those individuals. However, down-weighting these subjects in the traditional mediation analysis made very little difference  suggesting that the inclusion of subjects with fewer data points does not significantly affect the results (and certainly not the conclusions) of the study.When using a two stage modelling approach  one should appropriately propagate uncertainty in estimates from the first stage into the second stage model. In the traditional mediation analysis we weighted by the inverse of the approximate variance of the estimated growth outcomes to account for between-subject differences in their precision. While such weighting deals with relative variability in precision it does not allow for complete propagation of the first stage uncertainty, so the precision of the estimated effects may be somewhat overestimated in both analyses. Solutions include bootstrapping the whole process and joint modelling , but theIn this large-scale representative sample of the Scottish population we have elucidated the mediatory role of birth weight in the relationship between socioeconomic disadvantage and childhood growth in terms of height and overweight, and highlighted the importance of correctly accounting for intermediate confounding in mediation analyses. We found a strong, graded, negative effect of socioeconomic disadvantage on height which was mediated to some extent by birth weight and a strong, graded, positive direct effect of socioeconomic disadvantage on overweight which was partly masked by inconsistent mediation by birth weight.Our findings suggest that higher birth weight in more disadvantaged groups may be associated with reduced social inequalities in height but also with increased inequalities in overweight.S1 Appendix(PDF)Click here for additional data file.S2 Appendix(PDF)Click here for additional data file.S3 Appendix(PDF)Click here for additional data file.S1 Table(PDF)Click here for additional data file.S2 Table(PDF)Click here for additional data file.S3 Table(PDF)Click here for additional data file.S4 Table(PDF)Click here for additional data file.


Some etiological factors involved in developmental dysplasia of the hip (DDH) occur in the last trimester of pregnancy, which could result in a decreased incidence of DDH in preterm infants. The aim of this study was to compare the incidence of DDH between preterm and term infants.Ultrasound of the hip joint was performed in 2,534 term infants and 376 preterm infants within the population-based Survey of Neonates in Pomerania (SNiP) study.p < 0.023; 95% CI 0.012–0.022, χ2 test). A familial disposition for DDH was found in 169 (6.7%) term infants and 181 (7.1%) infants in the overall population. In preterm infants, dysplasia of the hip was found in only three late preterm infants with gestational age between 36 and 37 weeks (n = 97) and not in preterm infants <36 weeks gestational age (n = 279). Regression analysis revealed a narrowly significant association between gestational week of birth and DDH .A total of 42 (1.66%) term infants had DDH . Eighteen infants had bilateral findings. Hip dysplasia occurred more frequently in female neonates (32/1,182 vs. 10/1,302, Our study suggests that preterm infants <36 weeks gestational age have a decreased risk of DDH. Developmental dysplasia of the hip (DDH) is one of the most common congenital malformations with a regionally varying incidence of 1.5–4.9% –8. In 19Hip development is based on the interaction between genetically determined maturation of the cartilage and bone elements of the acetabulum and the pressure applied by the muscular forces of the centered femur head . RegardiOnly a few studies have reported the incidence of hip dysplasia in preterm infants, which has been reported to be 0.2–2.8 –13 compaIntrauterine risk factors for DDH with respect to less intrauterine space occur in the last trimester and are partially or completely absent in the case of preterm birth. The objective was to test the hypothesis that the incidence of DDH is lower in preterm infants than term infants by comparing the incidence of hip dysplasia between these groups. For this purpose, the population-based cohort study Survey of Neonates in Pomerania (SNiP) served as the database .The SNiP study is a prospective, population-based survey that collected comprehensive data on pregnancies, mothers, and their offspring in eastern Pomerania. Details were reported elsewhere . All neoTrained physicians participating in the survey according to a standardized protocol collected the data. The questionnaires were pseudonymized and entered into the study database by a medical archivist.In the study, data for 95% of all pregnant women in the Ostvorpommern region were recorded and >80% of mothers actively participated over the years. For the non-participants, risk factors during pregnancy, birth mode, and outcome of the child were recorded.SNiP received a positive vote from the Ethics Committee of the University Medicine Greifswald and written informed consent was obtained from the parents.Gestational age was defined as the time elapsed during the pregnancy from the first day of the woman’s last menstrual cycle to the current date. A normal pregnancy ranges from 37 to 42 weeks. To investigate the influence of gestational age on hip development, preterm infants were categorized as late preterm (34 weeks and 0 days–36 weeks and 6 days gestation), very preterm (28 weeks and 0 days–33 weeks and 6 days gestation), or extremely preterm (less than 28 weeks gestation). To investigate the influence of birth weight, preterm infants were categorized as having normal birth weight , low birth weight , very low birth weight , or extremely low birth weight .n = 2,910) were included in this subgroup-analysis of the SNiP-cohort. The subgroup represented 42.6% of all live births and 45.4% of all infants included in SNiP between January 2002 and November 2008  in the maternity ward. Briefly, the U2 comprises a physical exam, medical history, and counseling the mother on preventive measures, such as Vitamin D and K prophylaxis and vaccination. Preterm infants were screened at the corrected age of more than 36 weeks. Ultrasound scans, documentation, and classification were performed according to Graf . The staA familial disposition for DDH was defined as dysplasia of the hip in first degree family members , or family members with hip replacement before 50 years of age.n = 200) of performed hip ultrasounds. The final board certification implied an assessment of the candidate’s ability to perform an ultrasound. In addition, a DEGUM II certified examiner supervised all pathological results . All research was conducted in a Graf storage tray [All examiners passed their specialist’s examination as a pediatrician. The structured training comprised a minimum number  carried out all analyses using Stata 13.1 .Continuous data were expressed as mean and standard deviation according to their normal distribution and categorical data as absolute numbers and percentages. Differences in mature hip prevalence across groups of gestational age and birth weight were tested by Fisher’s exact test. We also performed a Poisson regression to test whether gestational week of birth was significantly associated with the prevalence of hip dysplasia. A p < 0.001) and more often born in breech presentation than term infants (p < 0.001). A familial disposition for hip disease did not differ between term and preterm neonates. Table Hip ultrasound was documented in 2,534 term neonates and 376 preterm neonates born between January 2004 and November 2008  and right hip (p < 0.001), preterm infants had a significantly higher incidence of mature hips than term infants. The incidence of Graf II was twice as high in term infants as in preterm infants. Overall, the incidence of Graf II c or higher dysplasia of the hip was greater in term infants  than in preterm infants . By individual years, the incidence of hip dysplasia in term infants was 2.10% in 2004, 2.23% in 2005, 2.2% in 2006, 0.9% in 2007, and 1.5% in 2008. A familial disposition for dysplasia of the hip was found in 169 (6.7%) term infants and 181 (7.1%) infants in the overall population.For the left (p < 0.001) and right hip (p < 0.001), extremely preterm and very preterm neonates had a significantly higher incidence of mature hips than late preterm babies (χ2 test). The incidence of Graf II a was 10-times higher in late preterm infants than in extremely preterm infants, and three-times higher than in very preterm neonates. Only three late preterm infants with a gestational age of more than 36 weeks had dysplasia of the hip: two female preterm neonates  and one male preterm neonate . Preterm infants with extremely low or very low birth weight had a higher incidence of mature hips than preterm babies with low or normal birth weight for the left (p < 0.001) and right hip (p = 0.004). The incidence of type II a was six-times higher in preterm infants of normal birth weight than preterm infants with extremely low birth weight. In the Poisson regression, we found no significant differences between term infants (1.75%) and preterm infants (0.53%) with respect to the prevalence of hip dysplasia . Therefore, we added analyses based on the continuous gestational week of birth variable, which revealed a barely significant association between gestational week of birth and dysplasia of the hip . One week of gestation was associated with a 16% increased risk of hip dysplasia , which may have influenced the results due to the low incidence of DDH. However, the strengths of this study were the large number of newborns, particularly preterm infants, investigated and that the cohort was population-based.Our study suggests that preterm infants  have a decreased risk of DDH. Furthermore, the incidence of fully mature hip joints was twice as high in preterm infants as in term infants and the proportion of mature hips was highest in the group with the most immature infants. These findings suggest that the absence of intra-uterine factors, such as lack of space, atypical positioning , and maternal steroid hormones, promotes maturation of the hip in preterm infants.


Though disparities in birth weight by race/ethnicity have been extensively reported in the United States, few studies have systematically investigated factors attributing to its variability. For 10,638,415 singleton infants born during 2009–2012 in the United States, we examined birth weight differences among 14 races and ethnicities , after sequentially adjusting for maternal, socio-economic and behavioral factors. Average birthweight of non-Hispanic white infants was 3381 g, while for other races/ethnicities birth weight ranged from being 289 g smaller in Japanese to 126 g larger in Samoan infants. Factors explaining differences of more than 50 grams in birth weight compared to white infants were: gestational age for black infants, height and body mass index for all Asian and Samoan mothers, and gestational weight gain for Japanese mothers. Difference in maternal age, parity, socioeconomic and behavioral characteristics did not account for significant portion of birthweight variations for any race. Our findings suggest that differences in maternal anthropometrics, gestational weight gain, and preterm birth rate, but not in maternal age, parity, socioeconomic or behavioral characteristics contribute to racial/ethnic differences in birthweight. Disparities in birthweight outcomes by race and ethnicity have been extensively reported12346The United States (US) is one of the most racially and ethnically diverse populations, with over 3 million infants of various races/ethnicities and national origins born each year. Numerous studies have found interracial differences in birth weight. For example, black and Asian infants are generally smaller than white infants1291012112141516913Recently the US national birth certificate database  to 1.3% (Indian). 15% had mixed race (parents were of different race or at least one parent was of mixed race), and 20% had at least one parent with race unknown.st trimester while 3.1% of Guamanians received no prenatal care; 2.3% of white women delivered outside of hospital compared to 0.1% of Samoans, and 66.1% of American Indians were Medicaid users compared to only 6.0% of Japanese.2) and obese (BMI > 30 kg/m2) mothers (89.0%) and Japanese had the largest number of underweight (BMI < 18.5 kg/m2) (19.4%) mothers. Samoans (167.5 cm) were tallest and Vietnamese shortest (156.7 cm). Koreans (47.9%) had the largest and Samoans (25.8%) the smallest proportion of null para mothers.Samoans had the largest proportion of overweight  and lowest among Japanese (mean 11.2 kg). Pre-pregnancy and gestational diabetes was most frequent in Samoans (9.2%) and lowest in Japanese (3.2%). Pre-pregnancy and gestational hypertension was most frequent in black mothers (8.0%) and lowest in Japanese (1.2%) mothers.Mean birth weight for non-hispanic white infants was 3381 g, and ranged from 3093 g in Japanese to 3507 g in Samoans. Only Hawaiians had more female infants than male infants. The preterm delivery rate was highest in black infants (14.2%) and lowest in Koreans (5.4%).As distribution of pre-pregnancy BMI differed by race/ethnicity, and gestational weight gain recommendations differ by BMI, we have shown the distribution of average gestational weight gain stratified by BMI and race/ethnicity in 2 = 0.38 in final model), and nearly 60% of the variability in birthweight due to race/ethnicity could also be explained by these factors .Next, in We also compared average birth weight z-scores limited to gestational weeks 34–41 across the 14 races/ethnicities shown in . AverageRegarding characteristics that correlated with racial difference in average birth weight by over 50 g, we noted the following: shorter gestational age for black infants [+81 g], lower maternal BMI and height for Asian infants , higher maternal BMI & height for Pacific Islanders (Guamanian [+68 g] and Samoans [−99 g]); and lower gestational weight gain for the Japanese (+60 g). Distribution of maternal age, parity and other socioeconomic and behavioral factors  barely added to explaining any difference in average birth weight for any race/ethnic group.To our knowledge, this is the first comprehensive study using a US population sample to explore parental characteristics associated with race/ethnic variations in birth weight. Background maternal characteristics including anthropometric, obstetric, socioeconomic, and behavioral factors varied widely. Factors that explained over 50 g of the difference in mean birth weight compared to whites included: gestational age for black infants, maternal height and BMI for all Asian and Samoan mothers, and gestational weight gain for Japanese mothers. Nevertheless, socioeconomic and behavioral factors failed to account for substantial (i.e. >50 g) differences in birthweight between race/ethnic groups.Our study highlights potential target areas for maternal interventions to reduce adverse birth outcomes. For example, gestational weight gain was markedly higher in Samoan women. Adult Samoans have a higher prevalence of overweight/obesity compared to other race/ethnic groups92223The preterm birth rate was highest among black mothers, which is consistent with previous reports18A study in the United Kingdom used a similar approach to examine the contribution of socioeconomic, maternal anthropometric and behavioral factors to race/ethnic differences in mean birth weightAlthough the specific minor races/ethnic groups within these two countries differ from the United States and are not directly comparable, both studies supported our findings that racial differences in maternal anthropometrics may be the largest driver in racial variability in fetal growth, and consideration of maternal size reduces the disparity in birth weight distribution across races/ethnicities.Currently a unified national birth weight reference is used to diagnose an infant as too small or too large for gestational age, both which are risks for unfavorable consequences272829303132333435We also found that the magnitude of explanation of birth weight variability by maternal factors differed by race/ethnicity. Of the 11 races/ethnic groups with over a 100 g difference in mean birth weight compared to white infants, our model explained over 80% of the difference for Vietnamese, Chinese, and Samoans, while over 60% of the difference remained unexplained for black and Asian Indian infants. Many previous studies failed to explain the birth weight difference between white and black infants, even after adjustment for maternal, social and behavioral factors184910111939Our study has several limitations. First, we did not have information on paternal anthropometrics. As we found maternal anthropometrics affect mean birth weight differences among races/ethnicities, paternal anthropometrics may be one of the factors causing the unexplained disparity among races/ethnic groups. Second, although we found that maternal anthropometrics, gestational weight gain and gestational length explained a substantial portion of racial differences in birth weight, as we were not able to differentiate US-born infants and non-US births, we do not know whether these characteristic differences are genetic, or due to cultural or environmental diversity. Third, although our study was based on national data, our database was limited to states that used the 2003 revision of the US birth certificate, hence our coverage was 66–86% of all births. Finally, the natality data is a combination of forms filled in by the mother as well as by staff of the birth facility, with no strict standardization of completeness or accuracy, and differences in completeness and accuracy between races may have caused bias. Increased standardization on by whom and how thoroughly these forms should be filled out, would improve accuracy in research results derived from natality data in future studies.Despite these limitations, this is the first and largest study to look at birth weight among races/ethnicities that investigates how maternal characteristics including anthropometrics explain variability in birth weight in the United States. Further study is needed to reveal whether such differences are inherited and thus not modifiable, or whether the unexplained racial differences reflect exposures that were not captured in the birth records, e.g. cumulative exposure to socioeconomic disadvantage across the life course, as well as the historical and ongoing legacy of race/ethnic discrimination of minority groups in society.For this study we used the National Natality FileData was restricted to those births recorded on the 2003 revised birth certificates. This restriction was applied because the reporting of multiple-race/ethnicity categories and details on maternal anthropometrics, such as maternal height, pre-pregnancy weight, delivery weight, are included in the 2003 revised version of the birth certificate but not the older 1989 version. In 2009, 28 of 50 states in the US were using the 2003 revised version of the birth certificate, and this increased to 38 plus Washington DC in 2012, representing 66% of all US births in 2009 and 86% in 2012.Of 12,375,027 singleton births within 2009–2012, the following subjects were excluded from analysis: missing information on birthweight , missing information on gestational age , delivered before 22 weeks or after 42 weeks , missing information on mother’s height or body mass index (BMI) , missing information on gestational weight gain , missing information on birth facility or initiation of prenatal care . Thus, analysis was done on the remaining 10,638,415 infants.The 2003 US birth certificate documents self-reported maternal and paternal race/ethnicity, and Hispanic origin. Race is recorded as either single or ‘bridged’ race/ethnicity, with ‘bridged’ meaning mixed race/ethnicity. Single racial or ethnic categories included white, black or African American, American Indian/Alaskan Native, Asian Indian, Chinese, Filipino, Japanese, Korean, Vietnamese, Native Hawaiian, Guamanian, Other Asian, or Other Pacific Islander. Hispanic origin is documented as Hispanic  or non-Hispanic.Using these variables, we classified both maternal and paternal race/ethnicity as follows: non-Hispanic white, non-Hispanic black, Mexican, Puerto Rican, Cuban, American Indian, Asian Indian, Chinese, Filipino, Guamanian, Japanese, Korean, Vietnamese, native Hawaiian. Other race/ethnic groups (such as “Other Asian” “South American”) or unknown race/ethnicity, non-white Hispanics, and bridged race/ethnicity were classified as “Mixed, Other or Unknown Race”.As our interest was mainly in infants of parents with a distinct race/ethnicity, we defined “infant race” as infants born to parents with a matching single race/ethnicity, or who were assigned their parents’ race/ethnicity. All other infants who had either at least one parent with “Mixed, Other or Unknown Race”, or parents of different race/ethnicities, were classified as “Mixed, Other or Unknown Race”.st trimester, 2nd trimester, 3rd trimester, none, unknown). Obstetric characteristics were also categorized when used in the multivariate analysis models: birth order , maternal age (in years), BMI , height (in cm), gestational age (in weeks) and gestational weight gain (in kgs).The following socioeconomic and behavioral factors were used: education , marital status , smoking status , medical insurance status , and initiation of prenatal care , and socioeconomic and behavioral factors . We did not use gestational hypertension and gestational diabetes because we considered them to be mediating factors rather than confounders.Birth weight z-scores were calculated using the US birth weight reference by Oken, which is based on natal data from births 1999–2000. Multivariate linear regression was used to calculate the association of race/ethnicity with birth weight and birth weight z-scores, after adjusting for potential confounders. Further, partial RThe protocol for this study was developed according to “Ethical Guidelines for Medical and Health Research Involving Human Subjects” and approved by the Institutional Review Board at the National Center for Child Health and Development, Tokyo, Japan (Project #2014-09). All descriptive and statistical analyses were performed using STATA version 13 . Statistical significance was set at the 0.05 level, and all statistical tests were two-tailed.How to cite this article: Morisaki, N. et al. Social and anthropometric factors explaining racial/ethnical differences in birth weight in the United States. Sci. Rep.7, 46657; doi: 10.1038/srep46657 (2017).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


The relationship between low iodine status and pregnancy-associated comorbidities has been known for decades. The relationship between iodine intake and placental pathologies is, however, far less clear. This study was designed to examine the relationship between dietary iodine intake and placental size while also focusing on typical adverse pregnancy outcomes.The dietary iodine intake of 4711 pregnant women enrolled in the Czech part of the European Longitudinal Study of Pregnancy and Childhood (ELSPAC) in 1990–1991 was established using a 145-item food frequency questionnaire. Multivariate linear regression models were used to estimate the relationship between dietary iodine intake during pregnancy and placental weight. Additional models were constructed to investigate the relationship between estimated dietary iodine intake and adverse birth outcomes.p = 0.011). Moreover, estimated dietary iodine intake was found to be significantly positively associated with the birth weight / placental weight ratio in the fully adjusted model .The estimated average iodine intake in the ELSPAC cohort was 106.6 μg/day. In the fully adjusted model, estimated dietary iodine intake was found to be significantly negatively associated with placental weight (β = −0.025, 95% CI: -0.044; −0.006, This study provides evidence of a relationship between estimated dietary iodine intake and placental weight and the birth weight / placental weight ratio. Additional research is warranted to provide more insight into the role of iodine in early as well as late placentation. Iodine is an essential micronutrient necessary for the synthesis of thyroid hormones. It has long been recognized that foetal as well as obstetric outcomes are dependent on normal maternal thyroid function . Severe Physiologically, gestation calls for significant alterations in the functioning of the thyroid system, with increased maternal iodine requirements consequently reaching over 50% , 4. ThisHuman chorionic gonadotropin (hCG), a pregnancy-associated hormone, is involved in the stimulation of the thyrotropin receptor (TSH-receptor), resulting in the consequent increase in the level of the circulating hormone leading to suppressed TSH levels via negative feedback. In addition to hCG, factors directly influencing maternal thyroid function include i) increased renal iodine clearance, ii) estrogen-dependent increase in thyroxine-binding globulin and iii) placental deiodinase activity. The establishment of cut-off values for thyroid hormones in pregnancy/lactation therefore requires an approach which takes into account gestational age. Among the above mentioned factors, increasing iodine clearance is of considerable importance in pregnancy as renal plasma flow increases by 75% in mid-pregnancy and there is a 50% higher glomerular filtration rate (GFR) between the first and third trimesters [As iodine is deeply involved in the development and maintenance of a healthy pregnancy, pregnant women and their foetuses as well as infants are most vulnerable to its deficiency. However, there is no unanimous consensus on the issue of iodine supplementation in pregnancy. In 2007, the WHO/UNICEF/ICCIDD  reached According to the WHO, the Czech Republic now ranks among countries where iodine deficiency is not considered a major public health issue. However, the Central European region has a long history of moderate to severe iodine deficiency. Salt iodisation was first introduced in Czechoslovakia (now the Czech Republic) in 1947, with a subsequent dramatic decrease in cases of severe iodine deficiency . Median While sufficient information on urine iodine levels is available, information on dietary iodine intake in the Czech Republic and its association with public health risk is very scarce. This study therefore aims to evaluate the estimated dietary iodine intake in pregnancy in a large prospective cohort from the European Longitudinal Study of Pregnancy and Childhood (ELSPAC), contextualize the achieved results with respect to dietary iodine and investigate the relationship between iodine intake and defined birth outcomes.ELSPAC is a population-based study designed to investigate environmental and other factors affecting the health and development of children from the prenatal period until adulthood . The stuEthical approval for the study was obtained from the European Longitudinal Study of Pregnancy and Childhood (ELSPAC) Law and Ethics committee and local research ethics committees. Written informed consent was obtained from all study participants and archived.Participating women were asked to answer two questionnaires during pregnancy, a questionnaire about themselves  and a questionnaire about their pregnancy (P2 – I am Expecting a Baby). A self-reported food frequency questionnaire (FFQ) was included as part of the P2 questionnaire package in accordance with the ELSPAC protocol  at 32 weFood estimates were combined with iodine content data from national databases and reported as μg of iodine per 100 g or 100 ml. To achieve a precise and complete evaluation of iodine intake from all possible sources, the following databases were used in hierarchical order: Czech Food Composition Database , Slovak Selected birth outcomes were treated as dependent variables in the models: birth length and weight, placental weight and cephalisation and ponderal indices. Gestational age, gestational diabetes, gestational hypertension, stillbirth, congenital malformations, pre-eclampsia and IVF pregnancy were included as independent variables. All data were collected by healthcare professionals at the maternity hospital at the time of birth. Gestational age was determined using ultrasound dating.Pre-eclampsia was defined in accordance with the International Society for the Study of Hypertension in Pregnancy (ISSHP) criteria  as systo3). The cephalisation index was expressed as the ratio of head circumference at birth (cm × 104) to birth weight (g) and later expressed as μm/g [Birth weight was measured to a precision of 0.1 kg. Placental weight was reported by the doctor present at birth. All possible gestational complications were summarised in the physician’s birth questionnaire (N1). The ponderal index was calculated as birth weight (kg) divided by birth length cubed , gestational age, maternal pre-conceptional weight, maternal age at birth, maternal smoking prior to pregnancy , living with partner status (yes–no), self-reported alcohol intake during pregnancy and self-reported drug abuse during pregnancy. Maternal socioeconomic status, a variable generally associated with adverse birth outcomes, was assessed based on several variables: maternal education (categorized according to ISCED), maternal job (categorized according to ISCO 88 and ISCO 08 into 16 groups and 4 skill levels), ethnicity  and family income calculated according to Eurostat recommendations for equalized family income , and appDietary iodine intake was analysed using 44 food items with established iodine content selected from FFQs administered to study subjects. Uniformity was tested by examining the distribution of variables in a loading plot verified using the Kaiser-Meyer-Olkin (KMO) measurement of adequacy with a value of 0.5 considered as acceptable for further analysis. The Bartlett sphericity test was used to establish whether the correlation matrix differed significantly from the identity matrix . OrthogoThe relationship between maternal dietary iodine intake and birth outcomes was investigated using linear regression models. Multivariate models were built for each birth outcome separately in two steps. First, the association between maternal dietary iodine intake and birth outcomes was investigated using a crude unadjusted analysis using the univariate models. Second, biological factors , social and demographic characteristics , maternal health and comorbidities of the mother and child , maternal alcohol consumption, substance abuse, smoking and use of nutritional supplements during pregnancy, and finally gestational age and sex of the child were included as potential confounders into final (adjusted) multivariate models.P-values of less than 0.05 were considered statistically significant.Missing data were imputed using the Multiple Imputation by Chained Equations method  and regrThe basic anthropometric and demographic characteristics of the study subjects are provided in Table 2 = 14,435, df = 630, p < 0.001) provided strong statistical evidence that the correlation matrix was significantly different from the identity matrix.Food group analysis was performed using factor analysis. The adequacy of data for analysis was evaluated based on the KMO value ; the resThree factors identified in pregnant women explained 14.6% of total variance. The first factor – which accounted for 6.2% of total variance among mothers – was labelled the prudent factor. This factor may be characterized by high factor loading in the case of whole-wheat bakery products, fish, legumes, fresh vegetables, root vegetables and rice. The second factor – which accounted for 4.5% of total variance – was labelled the carbohydrate–dairy factor and included foods such as refined flour bakery products, pudding, cheese, fresh fruit, milk and dairy products. The third factor – which accounted for 3.9% of total variance – was labelled the junk food factor and included items such as biscuits, wafers, pies and cakes, French fries, roasted potatoes, pâté and ground meat. The factors scores of the individual patterns have been reported in our previous paper focusing on the association of anthropometry with dietary patterns of mothers in the third trimester .In order to address the potential association between estimated iodine intake and placental size in the context of birth outcomes, we first performed an estimation of average dietary iodine intake by assigning average iodine intake to individual food items. Since no longitudinal comprehensive Czech database providing measured iodine levels for a broad spectrum of foods in the relevant period (1991–1992) is currently available, we derived iodine intake data from multiple available databases and systematically contextualized iodine content with individual FFQ items. As considerable differences in reported iodine intake between databases from various geographical regions were established, we based our methodology on the origin of iodine in foods in close geographical regions. For example, while the Czech Food Composition Database  reportedOnce various birth outcomes included in the initial model were reduced to dependent variables, we obtained a single final model with a statistically significant relationship between estimated iodine intake and placental weight. We observed a negative association which indicates that every 100 μg of daily iodine intake during pregnancy is significantly associated with a 2.5% decrease in placental weight .p = 0.011).Quartiles of cohort subjects with respect to estimated iodine intake are provided in Table The results of univariate as well as multivariate regression modelling for the logarithm of birth weight / placental weight ratio are provided in Table Multiple physiological changes in thyroid function take place during pregnancy; aberrant thyroid function during pregnancy has been associated with adverse obstetric outcomes . Since tBased on our results, iodine intake may potentially change the weight of the placenta or the birth weight / placental weight ratio by several percent. This supports a hypothesis which states that the placenta may serve to some extent as an iodine reservoir capable of expanding in size to increase its absorption potential in the case of lower dietary iodine intake, i.e. in a manner similar to the thyroid gland. The same association of iodine intake and placental size has been observed in mice . Based oThe above mentioned findings are in line with the current understanding of the role of the trophoblast in the regulation of iodine and thyroid hormone transfer from the mother to the foetus. While the foetal thyroid starts to concentrate iodine after the first 10–12 weeks of gestation, the synthesis and secretion of the thyroid hormone under TSH control starts at approximately 20 weeks of gestation . Prior tThe transplacental passage of thyroid hormones from mother to foetus and the following supply of the thyroid hormones into trophoblast cells requires the expression of placental plasma membrane transporters for thyroid hormones. The human placenta displays a wide range of transporters for thyroid hormones from early gestation: MCT8, MCT10, LAT1, LAT2, OATP1A2 and OATP4A1; the observed expression patterns are different between pregnancies with normal physiology and pregnancies affected with intrauterine growth restriction . These tThe delivery of the maternal thyroid hormones to the foetus therefore depends on i) the availability of the thyroid hormones in the maternal circulation and ii) the functional effectiveness of the transplacental transport systems. Current knowledge of thyroid hormone transporters not only strongly suggests that the placenta plays a role in iodine uptake, but also supports the hypothesis that the placenta may act as a storage capacity supplying iodine to the foetus gradually . This isThe exact mechanism and reason for decreasing placental size with increasing dietary iodine is, however, unknown and deserves further investigation. Although the importance of aberrant placental size is generally unclear, some evidence indicates that placental size may be associated with future developmental/cardiometabolic conditions. Some studies suggest that placental size is correlated with birth weight as well as with placental function . Under cThe gold standard for the estimation of iodine status in the general population today is urinary iodine concentration measurement . StudiesAt the time of data collection, recommended daily iodine intake was generally 50–100 μg/day with 150 μg/day recommended for pregnant women . The recThe major limitation of the study is that due to the design of the study  we were unable to carry out adjustments for urinary iodine. Due to the large number of subjects, it was necessary to use food frequency questionnaires which were not formally calibrated against diet diaries or biomarkers, i.e. a common approach in the early 1990s. Since the FFQ constitutes a methodological instrument designed to measure long-term dietary behaviour, it is prone to several methodological limitations, including a finite list of foods and food groups . FinallyOn the other hand, it must be mentioned that this study is the first to investigate dietary iodine intake in relation to the dietary patterns of pregnant women in the transitional period of the early 1990s in Czechoslovakia. Moreover, it also provides insight into the significant relationship between dietary estimated iodine intake and defined adverse birth outcomes. Additional key points of this study include the relatively large sample size of pregnant women and multiple adjustment for confounding factors, although the possibility of residual confounding can never be entirely ruled out. In addition, the observed dietary patterns are consistent with patterns expected in early 1990s Czechoslovakia, which may be demonstrated e.g. by the absence of typical “imported” patterns such as Mediterranean and vegetarian diets not typical at the time.This study reveals the possible influence of dietary iodine intake and its effect on placenta size, which influences the dynamics of intrauterine growth. These results contribute to the much-needed debate on iodine intake during the vulnerable child development period. Following adjustment for multiple confounders, including socio-economic status, our study found placental weight to be the only birth outcome significantly associated with iodine intake. While our research indicates that the placenta may be partially capable of compensating for iodine deficiency, more research into mechanistic relationship between iodine accumulation and placental growth is necessary. Based on our results, however, we encourage researchers to use placental weight as a possible interaction parameter with iodine supplementation in future studies.


FH+) or without such histories (FH−), we examined the reproducibility and reliability of global and regional measures of fractional anisotropy (FA) values, measured using the Enhancing Neuro Imaging Genetics Through Meta Analysis (ENIGMA)‐diffusion tensor imaging (DTI) protocol. Highly reliable measures are necessary to detect any subtle differences in brain development.In preparation for longitudinal analyses of white matter development in youths with family histories of substance use disorders , mean absolute differences (MAD), and intraclass correlations (ICC) for global and tract‐specific FA values.First, we analyzed reproducibility data in a sample of 12 healthy young adults (ages 20–28) imaged three times within a week. Next, we calculated the same metrics in data collected 1‐year apart in the sample of 68 DTI‐FA values and most of the white matter tracts, except for the corticospinal tract and the fornix in both adults and youths. There was no significant effect of FH‐group on reproducibility (p = .4). Reproducibility metrics were not significantly different between adolescents and adults . In post hoc analyses, the reproducibility metrics for regional FA values showed a strong positive correlation (r = .6) with the regional FA heritability measures previously reported by ENIGMA‐DTI.We found excellent reproducibility for whole‐brain ENIGMA‐DTI FA, positing it as viable analysis tools for longitudinal studies and other protocols that repeatedly assess white matter microstructure.Overall, this study demonstrated an excellent reproducibility of  The MAD is the absolute difference between first and second visit and provides assessments of the absolute difference between two datasets. The MCV is calculated as the standard deviation normalized by the average for two visits and provides a general assessment of variance in each dataset relative to the mean. The ICC is a form of correlation analysis that assesses the consistency of variability among subjects on two visits. Thus, using all three metrics enables a thorough assessment of reproducibility.These three metrics were applied to the global and regional FA values in 12 healthy young adults (ages 20–28 years) imaged three times within a week to estimate variability in regional FA values to examine the stability of the technique. The reproducibility assessed in the healthy adults, within whom no change is expected in the given interval, will help us interpret our longitudinal assessment in the adolescent population. To examine the variability between time points in the adolescent cohort, we also calculated MAD, MCV, and ICC.22.1N = 21/68 FH−/+) from an ongoing longitudinal study of adolescent development and substance use involvement in youths at elevated risk for substance use disorders  were imaged three times within 1 week using the same protocol to calculate an estimate of the variance in FA values over this time period. These estimates of variance were also examined in 89 children and adolescents  images, 64 isotropically distributed diffusion‐weighted directions, and axial slice orientation with 50 slices and no gaps. The number of directions, b0 images, and the magnitude of the b values were calculated using an optimization technique that maximizes the contrast to noise ratio based on the average diffusivity of the cerebral white matter and the T2 relaxation times  equipped with a multichannel head coil using a protocol described elsewhere 's FNIRT  analytic method .ENIGMA‐DTI (RRID:SCR_014649) protocols were used to extract whole‐brain and tract‐wise average FA values for both datasets. These protocols are detailed elsewhere . QA/QC checks consisted of inspecting the data, vector gradients, registration, and average skeleton projection distance. The data were visually inspected during MRI session to eliminate obvious image distortions due to severe motion. If excessive blurring, data distortions, or loss of SNR were observed during the scan acquisition, the study was interrupted, subjects were instructed to keep still, and the DTI sequence was repeated.Quality assurance and quality control QA/QC was conducted according to the ENIGMA‐DTI protocol  differences between individual transformation matrices and a unit matrix using the RMSDIFF program distributed with FSL. The identity matrix, where diagonal elements are set to 1 and all others set to 0, indicates the case of no difference indicating no interscan motion. The RMSDIFF program reports an RMS difference (in mm) between two matrices: larger values indicate greater differences between the alignment matrix and the identity matrix. This number is sensitive to both translations and rotations, providing a good estimate of the motion. The average estimate per subject is obtained by averaging RMS distance for individual volumes.The average projection distance is another QA measurement proposed by the ENIGMA‐DTI workgroup. This measure assesses the registration quality between individual images and the ENIGMA‐DTI template by calculating the average projection distance for the group‐wide skeleton. The skeleton projection step corrects for minor spatial mismatch between the centers of the track on the template versus the individual image. A higher projection distance may indicate problems with aligning individual brain to the template. The ENIGMA‐DTI protocol currently specifies no critical cutoff for the projection distance, and average projection distance for ENIGMA studies of 3.6 ± 0.2 mm. However, in this study, we repeated DTI scans for sessions with projection distances that exceeded 3.8 mm.2.6p = .05/18, Bonferroni correction threshold for multiple comparisons). For the healthy young adult cohort, reproducibility between visits for 18 tracts was assessed via mean coefficients of variation (MCV), mean absolute differences (MAD), and intraclass correlations (ICC). Calculations were performed using SPSS . To estimate differences in variance over one year, these same metrics were calculate in the FH+ and FH− cohorts, separately as well as in both groups combined. t‐tests were performed to assess differences between MCV and MAD between FH+ and FH− with significance set at p = .0028 (p = .05/18). t‐tests were also performed to ensure that the subjects in the FH+ and FH− groups were not significantly different in terms of age or time between scans. Chi‐squared tests were performed to ensure they were not significantly different based on measures of sex, race, or ethnicity. To assess the differences in reproducibility measures between all three groups (young adults and FH+ and FH− youths), a one‐way ANOVA was performed with significance set to p = .0028 to account for multiple comparisons. Additionally, we performed a linear correlation analysis between the measurements of reproducibility calculated in the cohort of adolescents and the tract‐based estimate of heritability published by ENIGMA‐DTI study  and FH+ and FH− subjects . Likewise, there were no significant differences in the average projection distance for adults  and FH+ and FH− subjects . Mean FA values for each tract for both the adults and youths are shown in Table S1. There was no significant correlation between average FA values and either average motion or projection distance . In the post hoc analysis for the 25 imaging sessions that failed the QA/QC measurements, the negative correlations between FA and average motion or projection distance were also nonsignificant .Demographics for young adult participants in the replication study and adolescents participating in the longitudinal study are detailed in Table Reproducibility measures for the 12 young adults are shown in Table r = .95). The splenium and ACR had the lowest MAD values and the fornix and corticospinal tracts having the highest MAD values .The mean coefficients of variation (MCV) for the average FA across the full skeleton in FH+ and FH− youths combined was 1.65% Table . Regionas Figure , top rows Figure . DespiteSimilar patterns of variance were observed when FH+ and FH− youths were examined separately . Linear correlation analyses for the tract‐wise MCV and MAD values between FH+ and FH− groups showed excellent agreement between tracts with r = .837 and .841. Likewise, linear regression analyses between adults and all adolescents (both FH+ and FH−) showed good agreement between tracts with r of .66 and .63, respectively (p = .01). The correlation between adults and adolescents (FH+ and FH−) was not significant . Correlation analyses of tract ICC metrics between groups showed significant, albeit lower correlation  for FH+ and FH−.One‐way ANOVAs performed on tract‐wise MCV, MAD, and ICC values showed no significant differences (h2) reported by ENIGMA‐DTI   Figure .4We examined reproducibility of FA values provided by the ENIGMA‐DTI protocol in 12 healthy adults, scanned three times within 1‐week period, where no changes in white matter microstructure were expected. Parallel analyses were performed in a large cohort of adolescents imaged twice, 1‐year apart, where only modest developmental changes in white matter microstructure were expected. ENIGMA‐DTI protocol is unique in that it was specifically developed and validated to provide homogenized measurements of DTI parameters across diverse population samples and imaging protocols for genetic imaging analyses  reflect differences in underlying white matter microstructure rather than driven by methodological issues. Furthermore, the high reproducibility of the adult population global and regional FA metrics indicates this protocol will likely be sensitive to changes occurring as a result of normal adolescent development as well as potential neuroanatomical changes that may develop as a result of problem substance use.Reproducibility measurements obtained by our study were predictive of the heritability values reported by ENIGMA‐DTI Working Group (Kochunov et al., 5In summary, the ENIGMA‐DTI protocol yields highly reproducible tract‐based FA values, indicating that this protocol is suitable for monitoring white matter microstructural changes over time in longitudinal samples using the same DTI acquisition protocol. In our ongoing work, we will be using this protocol to monitor white matter microstructural changes in our cohort of FH+ and FH− youths due to normal development as well as changes resulting from the initiation and progression of substance use. Click here for additional data file. Click here for additional data file. Click here for additional data file.


Echinococcus spp. Rupture of the biliary tract is an important and serious complication. It is often diagnosed on a suspicious clinical findings by radiologic imaging procedures. But, complicated hydatid cyst disease can show different radiological images, as they can simulate any disease. Here, we reported a case which diagnosed cholangiocellular carcinoma in primary diagnosis by ultrasonographic evaluation but diagnosed hydatid cyst in endoscopy evaluation. A 30-year-old female patient was admitted to the hospital with 1 year continuous severe abdominal pain aggravated for 1 month.Liver cyst hydatic (LCH) is an important parasitic disease caused by Kefeli A, Basyigit S, Yeniova AO, Nazligul Y. Complicated Hydatid Cyst: Ultrasonographic Illusion and Endoscopic Diagnosis. Euroasian J Hepato-Gastroenterol 2015;5(1):65-66. To the Editor,Echinococcus spp. Rupture of the biliary tract is an important and serious complication. It is often diagnosed on a suspicious clinical findings by radiologic imaging procedures. But, complicated hydatid cyst disease can show different radiological images, as they can simulate any disease. Here, we reported a case which diagnosed cholangiocellular carcinoma in primary diagnosis by ultrasonographic evaluation but diagnosed hydatid cyst in endoscopy evaluation. A 30-year-old female patient was admitted to the hospital with 1 year continuous severe abdominal pain aggravated for 1 month. In physical examination, there is no positive clinical sign other than sensitivity of right abdominal quadrant. In her laboratory findings, it was found that 9500/ul of total leukocyte count, 11.7 gm/dl of hemoglobin, 270000/ul of platelet, 2.3 mg/dl of total bilirubin (T Bil), 1.5 mg/dl of direct bilirubin (D Bil), 668 IU/l of alkaline phosphatase (ALP), 207 IU/l of alanine aminotransferase (ALT), 357 IU/l of as partate aminotransferase (AST), 512 IU/l of gamma-glutamyl transferase (GGT). Abdominal ultrasonography showed a 10 × 9 cm size mass and a 93 × 36 mm size dense content cyst including solid component inside the mass which fills nearly whole right hepatic lobe. Dilatation of intrahepatic biliary tree was detected and common bile duct was 19 mm in diameter and a solid lesion inside the distal section of the lumen appeared like cholangiocellular carcinoma. Meanwhile, patients developed overt jaundice, total bilirubin increased to 10.8 mg/dl, body temperature was measured as 39 to 39.5°C. One gram of ceftriaxone was given twice daily. The patient underwent gastroscopy; a lesion colored yellow-white was thought to be hydatid cyst membrane inside Vater’s papilla extending into the duodenum (Liver cyst hydatic (LCH) is an important parasitic disease caused by duodenum .Endoscopic retrograde cholangiopancreatography (ERCP) was made due to increase in bilirubin level and endoscopic evidence of the hydatid cyst membrane. Amounts of cyst membrane and materials of cyst was removed by using biliary occlusion balloon. Biliary tract of right lobe cannot be visualized. Nasobiliary drain was placed in the right intrahepatic bile ducts. Agglutination tests for hydatid cyst was positive by 1/320 and 400 mg of albendazole was given twice a daily. By control ultrasound, it was observed to be an 85 × 75 mm hypoechoic echogenicity with dense content suggesting that cystic lesion, and echogenicity belongs to the air in the intra- and extrahepatic bile ducts in the right lobe. Laboratory findings showed 2.0 mg/dl of T Bil, 51 IU/l of ALT, 61 IU/l of AST, 420 IU/l of ALP. Abdominal pain and jaundice of the patient was decreased. She was discharged with a treatment of 400 mg of albendazole twice a daily. The patient had no symptoms. The ultrasono-graphical assessment showed that she was normal after 3 months.Echinococcus spp. It is usually transmitted during childhood and remains silent lifelong. Jaundice with fever or even anaphylactic reactions may occur, if it is complicated. The most important complications of LCH are the infection of the cyst, obstructive jaundice due to compression or rupture of intrahepatic bile ducts, rupture through peritoneum, or the spread to other organs. Rupture of the biliary tract is an important and serious complication. It is often diagnosed on a suspicious clinical findings by radiologic imaging procedures.1Liver cyst hydatic is an important parasitic disease caused by Complicated hydatid cyst disease can show different radiological images, as they can simulate any disease. In our case, the sonographic examination of the 10 cm mass insisted to consider cholangiocellular carcinoma as a primary diagnosis. Endoscopic evaluation was made due to increased dyspepsia and overt jaundice on the 3rd day of admission and presence of yellow lesions extending into the duodenum.It is uncommon to make diagnosis of intrabiliary rupture of LCH by regular endoscopic examination. Endoscopist should recognize the appearance of cyst membrane protruding into duodenum from Vater’s papilla and avoid removing it because of high inoculation risk. In this article, we reported the endoscopic appearance of cyst membrane of an intrabiliary rupture of LCH in the second part of the duodenum.


During the last decade, a rapid increase of birth locations for low-risk births, other than conventional obstetric units, has been seen in the Netherlands. Internationally some of such locations are called birth centres. The varying international definitions for birth centres are not directly applicable for use within the Dutch obstetric system. A standard definition for a birth centre in the Netherlands is lacking. This study aimed to develop a definition of birth centres for use in the Netherlands, to identify these centres and to describe their characteristics.International definitions of birth centres were analysed to find common descriptions. In July 2013 the Dutch Birth Centre Questionnaire was sent to 46 selected Dutch birth locations that might qualify as birth centre. Questions included: location, reason for establishment, women served, philosophies, facilities that support physiological birth, hotel-facilities, management, environment and transfer procedures in case of referral. Birth centres were visited to confirm the findings from the Dutch Birth Centre Questionnaire and to measure distance and time in case of referral to obstetric care.n = 3), alongside (n = 14) and on-site (n = 6). Transfer in case of referral was necessary for all freestanding and alongside birth centres. Birth centres varied in their reason for establishment and their characteristics.From all 46 birth locations the questionnaires were received. Based on this information a Dutch definition of a birth centre was constructed. This definition reads: “Birth centres are midwifery-managed locations that offer care to low risk women during labour and birth. They have a homelike environment and provide facilities to support physiological birth. Community midwives take primary professional responsibility for care. In case of referral the obstetric caregiver takes over the professional responsibility of care.” Of the 46 selected birth locations 23 fulfilled this definition. Three types of birth centres were distinguished based on their location in relation to the nearest obstetric unit: freestanding (Twenty-three Dutch birth centres were identified and divided into three different types based on location according to the situation in September 2013. Birth centres differed in their reason for establishment, facilities, philosophies, staffing and service delivery. Throughout the world, birth centres are regarded as homelike settings where women with uncomplicated pregnancies can give birth with a midwife with the assistance of a maternity care assistant (MCA). When complications arise or when medicinal pain relief is requested, referral to a hospital obstetric unit takes place –5. BirthDutch women, considered at the start of labour to have low obstetric risk, can choose the place where they want to give birth: at home or out of home. Out of home birth can take place within a hospital setting or in a birth location outside of a hospital. The woman’s own community midwife is the responsible caregiver during labour and birth, regardless the location. She works autonomous and independent in a local midwifery practice. To work as a midwife in the Netherlands four years of education at the midwifery academy (Bachelor) have to be completed. After that, you are obliged to register in a nationwide register for health professionals . Dutch mDuring the last decade, a rapid increase in the number of out of home birth locations has been seen in the Netherlands. Several factors may be responsible for this sudden increase: women’s choice for home birth has decreased in recent years, leading to a higher demand for alternative birth locations that could not be provided by hospitals . BesidesThe varying international definitions for birth centre are not directly applicable for use within the Dutch obstetric system where the place of birth is interrelated with the clear role division between primary and secondary obstetric care.Between 2013 and 2016 the Dutch Birth Centre Study was carried out to evaluate birth centre care provision and its effects on perinatal outcomes, experiences of clients and caregivers and economic outcomes . This evThis study is part of the Dutch Birth Centre Study and aimed to develop a standard definition of birth centre for use in the Netherlands in order to identify all Dutch birth centres and to describe their characteristics.The methods used in the development of the birth centre definition were 1) the primary data collection, 2) a literature review and 3) a consensus process.Three different data collection tools were used. The first one was a short digital survey to make a basic selection of potential birth centres in the Netherlands. The second one was the Dutch Birth Centre Questionnaire, used to get more information about the characteristics of these presumed birth centres and the third tool was the semi-structured interview for the confirmation and elucidation of earlier findings.This tool was developed to obtain information about the place of birth options for low obstetric risk women in the Netherlands. It enquired about the existence of a) a homelike location for birth services for b) low risk women, that c) differed from the conventional hospital labour and birth setting. It was sent to the chair of every group of obstetricians associated with each of the 98 hospitals with maternity care in the Netherlands and to the chair of the local midwifery peer group in the vicinity of each of those hospitals.A measurement tool for use in the Netherlands was developed based on an Australian questionnaire used to study birth centres . Permission was obtained for this survey tool that contains questions regarding issues as staffing, founding philosophies and physical characteristics of birth centres. Additional questions were added relating specifically to birth centre care provision in the Netherlands. These covered issues as initiators, reason for establishment, estimated number of births in 2013, need for transfer in case of urgent referrals and judicial status. The DBCQ consisted of 150 questions and was used to collect data from birthing locations that were presumed to be birth centres. In January 2014 all selected birth centres were asked to provide the number of actual births that took place at the birth centre in 2013.Semi-structured interviews were designed to gather information from directing managers of those birth locations that qualified as presumed birth centre. Topics addressed included aspects of management and clinical leadership. During these interviews, information received from the DBCQ was confirmed and additional information was collected regarding time and distance from the birth centre to the hospital obstetric unit. Depending on the local situation, the distance from the birth centre to the obstetric unit was measured by counting steps or by kilometres on a navigation system. Time for transfer by bed or car was measured using a stopwatch during a simulated referral with transfer situation. All interviews were conducted by one researcher (IB).In March 2013, international definitions of birth centres were searched in Pubmed and common elements within these definitions were identified. Using literature and data from the DBCQ, the characteristics of these elements were identified for the definition. A concept definition for birth centre was developed and discussed with the Dutch Birth Centre Study research group. Members of this group included 2 professors of obstetrics, 4 senior researchers and 3 PhD-students, two of whom were midwives (one practising).In addition, the Dutch Birth Centre Study Advisory Committee discussed and adjusted the concept definition until consensus was reached . After aBetween April 2013 and June 2013, the locations that might qualify as a birth centre were collected in collaboration with The Royal Dutch Organisation of Midwives (KNOV), College of Perinatal Care (CPZ) and STBN . A call was also posted in the popular LinkedIn Group “Dutch birth care in motion” to obtain information about other potential birth centres.The Short Digital Survey was sent to midwives and obstetricians working in the vicinity of the identified potential birth centres. If they responded positively for all three questions, the location was presumed to be a birth centre. This resulted in a list of presumed birth centres for the study.Representatives from each presumed birth centre were contacted by telephone, informed about the study and asked to participate. The local manager of each birth location was the primary person asked to answer the DBCQ. If the local manager was not available, the Chair of the Board or a midwife associated with the birth location was asked to respond on behalf of the birth centre. In July 2013, the DBCQ was sent by email to all presumed birth centres. Non-responders were contacted again in August 2013. All answers to the open-ended questions were analysed by two researchers (MHe and IB) and categorized after consensus was reached.The semi-structured interviews with managers of the presumed birth centres were conducted by one researcher (IB) between January 2014 and April 2015. In May 2015 all birth centres were identified made in line with the Dutch definition of a birth centre and based on the information from September 2013.Descriptive data analyses were conducted using the Statistical Package for Social Sciences (SPSS) version 22.0 .n = 35), did not differ from the conventional labour ward on the obstetric unit (n = 27) or were not accessible as a birth location for low risk women who start labour under care of a community midwife (n = 8). More reasons for exclusion could be appropriate for one birth location. The remaining 46 locations were considered to be presumed birth centres and received the DBCQ. All questionnaires were returned of which 44 were fully completed. Two questionnaires were returned incomplete because the questions were not applicable for these two birth locations as being a presumed birth centre.In total, 93 birth locations were identified as potential birth centres. After completion of the short digital survey, 47 birth locations were excluded because they were not homelike  population to be served, 2) responsible professional for care provided, 3) environment, 4) philosophy, 5) location in relation to the nearest obstetric unit, 6) need for transfer in case of referral and 7) management structure (midwife/obstetrician). Using the information from the DBCQ  the owner of the birth location; 2) the floor manager of the birth location; 3) a member of the board of the birth location; 4) a member of the board of an integrated organization in which the birth location is a participant or 5) participating in a committee which is responsible for the local care content of the birth location.Birth centres are midwifery-managed locations that offer care to low risk women during labour and birth. They have a homelike environment and provide facilities to support physiological birth. Independent community midwives take primary professional responsibility for care. In case of referral the secondary caregiver (obstetrician or paediatrician) takes over the professional responsibility of care.The following definition of a birth centre was developed Fig. :Birth cefreestanding birth centre is located separate from a hospital with obstetric services. In case of referral the woman needs to be transferred to a hospital with obstetric services which will normally be by car or ambulance.A alongside birth centre is located in a hospital with obstetric services or on such a hospital’s grounds, but separate from the obstetric unit. In case of referral the woman needs to be transferred which will normally be by bed or wheelchair.An on-site birth centre is located within an obstetric unit of a hospital. In case of referral the woman does not need to be transferred: the secondary caregiver (obstetrician or paediatrician) will enter the birthing room.An Three types of birth centres were identified based on location:Nineteen of the 46 presumed birth centres were excluded because they were not midwifery managed see Fig. . Twenty n = 21) mentioned more than one reason for establishment. The most stated reasons were: the wish for a more homelike environment as opposed to conventional birthing rooms within the obstetric unit (74%), and the possibility to provide one-to-one support during early labour (57%). Competition and marketing were also mentioned as reasons; Ten birth centres (44%) were opened in order to compete with other hospitals offering a birth location for women with low obstetrical risk. Birth centres also mentioned logistics as a reason for establishment: in two regions (9%) the distance to a referral obstetric unit was perceived as being too large without the establishment of a strategically placed birth location for low obstetric risk women. Seven birth centres (30%) reported establishment because of a capacity problem in hospitals or in primary care services . More than three quarter (78%) of the birth centres reported that local community midwives were responsible for initiating the establishment of the birth centre.Most of the birth centres (Table n = 9), resuscitation of the neonate (n = 8), postpartum haemorrhage (n = 4), (eclamptic) insult (n = 4), Apgar score below 7 after 5 min (n = 4), placental retention (n = 3), prolapse of the umbilical cord (n = 3) and foetal distress (n = 2). In those situations the secondary caregiver came to the birth centre in case of referral. In five of the 14 hospitals with an alongside birth centre there was also the possibility for low risk women to give birth under the care of their own community midwife on the conventional labour ward. The rooms on this ward were different in environment, staffing, service and facilities compared to the rooms in the birth centre. Transfer time from the alongside birth centre to the nearest obstetric unit varied between 10 s and 3.5 min.Fourteen birth centres were located in a hospital but separate from its obstetric unit . In six of these birth centres referral with transfer to secondary care meant a move to another floor by elevator. Exceptions for transfer were locally described and included situations as shoulder dystocia (Six birth centres were located within an obstetric unit (on-site). For low risk women who gave birth at an on-site birth centre transfer was not needed in case of referral because the obstetrician with the obstetric team entered the room. Besides the other conditions as noted in Fig. All birth centres had a non-clinical homelike atmosphere. 74% of the birth centres have no medical equipment like a cardiotocography machine or a resuscitation bag and mask in sight. At the other birth centres this equipment in sight was minimalized by putting it not in a front position. All birth centres provided facilities to support pushing in a non-supine position , methods for discomfort and pain management that were allowed to be used in primary care (bath and shower) and one-to-one or one-to-two support by a maternity care assistant (MCA) as much as wanted and needed by the woman in labour and her partner.In all birth centres a MCA assisted the community midwife during labour, birth and postpartum. The MCA was part of the staffing of the birth centre in thirteen out of twenty three birth centres (57%). In twelve of these birth centres the MCA was 24/7 present. When not part of the staffing the MCA was on call for assistance during labour and came to the birth centre after a request by the community midwife. Midwives were not part of the staffing of the birth centre itself but were independent workers or part of the staffing of the larger organization that included the birth centre. They arrived at the birth centre only with a woman in labour or for postpartum care if applicable.In thirteen birth centres (57%) it was possible for the woman to stay for up to 10 days postpartum. In four of these centres the woman stayed in the same room as where she gave birth; in the other she had to change rooms on the ward or in the building. In all except one of these thirteen birth centres it was possible for the partner to stay one or more nights as well if desired. During the postpartum stay, a maternity care assistant was available on the ward 24 h per day in every birth centre. Hotel-like facilities were present in all 23 birth centres.Philosophies were ranked each from ‘not important’ to ‘very important’. The number of birth centres that ranked a philosophy as important or very important on the five point Likert scale are shown in Table The establishment of the birth centres was financed in many different ways. In 55% the local hospital was involved, in 32% a maternity care assistance organization, in 23% an insurance company, in 23% STBN and in 14% the community midwives. For two locations this information was unknown by the person who filled out the questionnaire. In 61% the birth centre itself was an independent legal entity.This study was undertaken to better understand the phenomenon ‘birth centre’ in the Netherlands. A standard definition for birth centre was developed, 23 birth centres were identified and their characteristics were described. Based on their location in relation to the nearest hospital obstetric unit, three different types of birth centres were seen: freestanding, alongside and on-site. Dutch birth centres differed in their reasons for establishment, services provided, founding philosophies, staffing and service delivery.In the Netherlands, the term ‘birth centre’ has a broad scale of meanings, varying from midwifery practices to locations for preconception consults, which is confusing –16. To hThis is the first study in the Netherlands that looked into the classification and description of the characteristics of birth centres. With this classification, it will be possible to study the effects of birth centre care provision on many different aspects such as perinatal outcomes and client and healthcare provider satisfaction . The intAll Dutch birth centres claimed that it was important to be committed to a physiological way of birth. We found that at on-site birth centres medical equipment was more often in sight than in alongside or freestanding birth centres. In addition, as on-site birth centres are located on the obstetric unit, there is easy access to technology and medical interventions during labour and birth. Physicians working at the obstetric unit are trained to look for pathology, and maybe therefore more likely to intervene. Stark et al. found that the support of physiological labour and birth for low risk women when giving birth at the obstetric unit is more difficult than at another location different from the hospital obstetric unit . TherefoBirth centres are homelike by having decorative changes like a specially designed bed and dim lighting and by providing hotel-like facilities. Facilities like a bath provide an option for non-pharmacological pain management that is associated with a significant reduction in risk of transfer and fewer interventions during labour , 22. A bWorldwide there is discussion about safety and distance of travel time from a freestanding birth centre to a hospital with an obstetric unit in case of referral during birth –29. TravIt was possible to develop a comprehensive definition for a Dutch birth centre that is based on the common elements found in international definitions with context specific characteristics for the Netherlands. From the many locations calling themselves birth centres, it was possible to identify and select birth centres in line with our definition. This methodology has contributed to the ongoing research into the effects of birth centre care provision and could be valuable for future research in this area.