Previous epidemiologic studies suggest associations between preterm birth and ambient air pollution.We investigated associations between 11 ambient air pollutants, estimated by combining Community Multiscale Air Quality model (CMAQ) simulations with measurements from stationary monitors, and risk of preterm birth (< 37 weeks of gestation) in the U.S. state of Georgia.Birth records for singleton births ≥ 27 weeks of gestation with complete covariate information and estimated dates of conception between 1 January 2002 and 28 February 2006 were obtained from the Office of Health Indicators for Planning, Georgia Department of Public Health . Daily pollutant concentrations at 12-km resolution were estimated for 11 ambient air pollutants. We used logistic regression with county-level fixed effects to estimate associations between preterm birth and average pollutant concentrations during the first and second trimester. Discrete-time survival models were used to estimate third-trimester and total pregnancy associations. Effect modification was investigated by maternal education, race, census tract poverty level, and county-level urbanicity.Trimester-specific and total pregnancy associations (p < 0.05) were observed for several pollutants. All the traffic-related pollutants  were associated with preterm birth . Associations tended to be higher for mothers with low educational attainment and African American mothers.Several ambient air pollutants were associated with preterm birth; associations were observed in all exposure windows.Hao H, Chang HH, Holmes HA, Mulholland JA, Klein M, Darrow LA, Strickland MJ. 2016. Air pollution and preterm birth in the U.S. state of Georgia (2002–2006): associations with concentrations of 11 ambient air pollutants estimated by combining Community Multiscale Air Quality Model (CMAQ) simulations with stationary monitor measurements. Environ Health Perspect 124:875–880; http://dx.doi.org/10.1289/ehp.1409651 Preterm birth is a well-known predictor of infant morbidity and mortality as well as increased long-term morbidity in adulthood . Several2), which are known to be spatially heterogeneous . Gestational age was defined according to date of last menstrual period, and date of conception was estimated by adding 14 days to the date of last menstrual period. By limiting to births ≥ 27 weeks we ensured that all births were followed for the entirety of the first and second trimesters. Preterm birth was defined as a birth < 37 weeks of gestation. We further excluded births where a) maternal residential address at delivery was unsuccessfully geocoded to the 2000 census block group , b) birth weight < 400 grams , c) mother’s age < 15 years or > 44 years , d) one or more identified congenital anomaly , and e) preterm births that had a procedure code for induction of labor . The final data set contained 511,658 pregnancies. Urbanicity for each pregnancy was assigned using a dichotomous variable based on the U.S. National Center for Health Statistics classification scheme according to county of maternal residence  and the frequency of measurements also varied across pollutants. Observations provide reliable temporal trends at and near monitors, and CMAQ simulations provide spatially rich information that is less reliable temporally. Our approach to combining these data involves three steps , 8-hr maximum ozone (O3), 24-hr average particulate matter ≤ 10 and ≤ 2.5 μm in diameter , and the PM2.5 components sulfate (SO42–), nitrate (NO3–), ammonium (NH4+), elemental carbon (EC), and organic carbon (OC). Pollutant concentrations for each pregnancy were assigned by linking the geocoded maternal census block centroid at delivery to the associated 12-km by 12-km grid cell based on the block group centroid. Ambient air pollutant concentrations were estimated by fusing 12-km by 12-km outputs from the U.S. EPA’s CMAQ model with ambient air pollutant concentration measurements from Air Quality System stationary monitors. The number of monitors ranged from 5 for CO to 42 for PMee steps . First, To evaluate model performance we created scatterplots to compare the daily pollutant measurements at the monitors with the estimated concentrations at the corresponding grids, with the relationships summarized using Pearson correlation coefficients. Leave-one-monitor-out cross-validation was used to estimate the Pearson correlation coefficient between the estimates and the measurements from the monitor that was held out of the model fitting. Fractional error—the ratio of the error of the estimate to the measured concentration—was calculated for each monitor in the leave-one-monitor-out cross-validation, and these fractional errors were averaged to calculate the mean fractional error.We used logistic regression to estimate associations between preterm birth (< 37 weeks of gestation) and average pollutant concentrations during the first trimester  and second trimester (weeks 14–26 of gestation) (model 1), and a discrete-time survival model to estimate associations during the third trimester and total pregnancy (model 2) to preclude bias due to differing exposure averaging periods between preterm and full-term births . SpecifiiY in model 1 is the dichotomous outcome indicating whether pregnancy i ended in a preterm birth , and itY in model 2 indicates whether a birth occurred during gestational week t for pregnancy i. The βi parameter is the estimate of the association with average pollutant concentrations in both model 1 and 2. In model 1, iX is the average pollution exposure during the first or second trimester for pregnancy i, and in model 2 itX is the time-varying average pollution exposure for pregnancy i between week 27 and the end of week t for the third-trimester model or between week 1 and the end of week t for the total pregnancy model. δ is a vector of parameters for the individual-level covariates (iZ) for pregnancy i. ϑ is a vector of fixed effects for the 159 counties, and γ in model 2 allows the baseline hazard to vary by gestational week.where We controlled for maternal education , race , smoking (dichotomous variable for self-reported tobacco use during pregnancy), and long-term trend using a natural cubic spline on conception date with 5 degrees of freedom  to account for the gradual increasing rate of preterm birth and decreasing concentrations of pollutants during the study period . Data onOdds ratios (ORs) are presented per interquartile-range (IQR) increase in pollutant concentration. Because the estimated association between air pollution and preterm birth is largely driven by within-county contrasts (due to the county-level fixed effects in the regression model), and given that each county has its own IQR for a given pollutant, we set the overall IQRs used for presenting the health associations as being equal to the median of the 159 county-specific IQRs. Average pollutant levels during pregnancy were used to calculate the IQRs, and we used the same IQR values to present results from the first-, second-, and third-trimester and total pregnancy models.p-values < 0.0001 for the differences among all categories of each characteristic)  of whom were preterm. Compared with mothers of full-term births, mothers of preterm births were more likely to reside in a large metropolitan county, be of African-American race, have a low education level, use tobacco products, and live in more impoverished census tracts  . Statewi3 was predicted best  whereas 1-hr maximum SO2 was predicted most poorly due to limitations in modeling of coal combustion plumes . Compared to the unfused CMAQ estimates, the predictions from the CMAQ data fusion model resulted in improved cross-validation Pearson correlation coefficients and decreased mean fractional errors for all 11 pollutants considered. When data were not withheld, the Pearson’s correlations between daily measurements at monitoring stations and fused CMAQ estimates at grids where monitors are located were all ≥ 0.90 except for PM10 (0.87) and SO2 (0.72) .Maternal residence census blocks were linked with 942 CMAQ grid cells, and the average distance between maternal census block centroid and the CMAQ grid cell centroid was 4.6 km . Cross-v2 1.81 ppb; SO2 1.59 ppb; O3 6.43 ppb; PM10 3.96 μg/m3; PM2.5 2.01 μg/m3; SO42– 1.27 μg/m3; NO3– 0.25 μg/m3; NH4+ 0.24 μg/m3; EC 0.14 μg/m3; and OC 0.36 μg/m3.  Correlations between average pollutant levels during pregnancy varied in magnitude, with correlation coefficients ranging from 0.03 (PM10 and NO2) to 0.95 (CO and NO2) (see Table S1).The distributions of daily pollutant concentrations are presented in p < 0.05) of preterm birth with IQR increase in first-trimester CO , NO2 , and SO2  were observed. All but three pollutants  were significantly associated with preterm birth during the second trimester, and in the third trimester CO, NO2, SO2, EC, and OC were significantly associated with increased risk of preterm birth. Associations for the total pregnancy period were frequently in the same direction as the trimester-specific estimates  for the risk of preterm birth per IQR increase in each of the 11 pollutants during the first, second, and third trimester and the full pregnancy are presented in stimates .2, and EC) as well as SO2  were frequently higher for mothers living in urban counties, for African-American mothers, and for mothers with lower education levels; however, the associations across the total pregnancy period were sometimes higher for mothers living in less urban counties. There was little suggestion of effect modification by percent of the census tract below poverty (results not shown).2, and EC changed little across the sensitivity analyses, whereas the point estimates for SO2 increased with additional control. For secondary pollutants that have a more pronounced seasonal pattern, such as O3 and SO4, and for PM10 and PM2.5, which are a mix of primary and secondary particles, the confidence interval widths increased with additional control for time. Because we controlled for county via stratification, our analysis was largely informed by contrasts in pollutant concentrations over time, so a regression spline that controlled for season likely decreased the exposure variability that informed the estimation of the pollutant associations for these pollutants that have seasonal patterns.Sensitivity analyses were performed by adding additional control to the regression spline on conception date (see Table S4). Associations with primary pollutants CO, NO2, and EC—and increased risk of preterm birth across all exposure windows. Several previous studies have reported higher risks of preterm birth associated with higher concentrations of CO during pregnancy .In addition to the associations with traffic-related pollutants, we observed statistically significant positive associations with SO2.5 components and preterm birth, although associations with OC and EC (averaged throughout the entire pregnancy) were reported in a Southern California study of preterm birth  with PM10, PM2.5, SO42–, and NH4+; however, given that the trimester-average pairwise correlations among these pollutants ranged between 0.64 and 0.93 (see Table S1), it is not surprising that the ORs for these pollutants were similar. These second-trimester findings contrast with results recently published by 3, which seems consistent with many previous findings; 3 and risk of preterm birth  and maternal education might modify relations between air pollution and birth outcomes in urban areas . In our Our study has several strengths. Foremost is the use of an innovative air quality model developed by combining CMAQ outputs with measurements from stationary monitors. This model provides complete spatial and temporal coverage over the study domain while reducing bias present in the CMAQ outputs . Cross-vSecond, the analysis is based on a population of > 500,000 births, and the time-to-event statistical approach used to estimate third-trimester associations avoids the potential bias resulting from differences in the length of the third-trimester averaging period between preterm and full-term births . For exa2, EC, and SO2) and smaller errors for secondary pollutants that tend to be more homogeneously distributed across space (such as O3 or SO42–). Even so, we tended to observe the most consistent evidence for associations in our study with the traffic pollutants , suggesting that if these measurement errors did cause bias to the null they were not so severe as to completely obfuscate our results. Maternal residential mobility is also a concern, because residence at the time of delivery was used to assign pollutant levels throughout pregnancy. A previous study conducted in Atlanta during 1993–1997 found that 22% of women moved residence during pregnancy, with approximately half moving within the same county  PDFClick here for additional data file.


To the Editor: The Global Polio Laboratory Network (GPLN) and the World Health Organization’s Polio Eradication Initiative (GPEI) accord high priority to detecting all vaccine-derived polioviruses (VDPVs) because they are neurovirulent and have the potential to cause outbreaks of poliomyelitis and establish poliovirus circulation. In patients with immunodeficiency diseases, persistent infections may become established with live oral poliovirus vaccine (OPV) and develop into VDPVs (>6 nucleotide substitutions (Sabin2) or ≥10 nucleotide substitutions (Sabin1 and Sabin3) in the VP1 region. The VDPV assays were found to be 100% specific for all 3 poliovirus types, 100% sensitive for Sabin1 and Sabin3, and 76% sensitive for Sabin2 —that is, withdrawal of Sabin2—beginning in April 2016  on January 2008. Sabin3 was isolated from stool sample 1 (R46064), collected 11 days after onset of paralysis. Stool sample 2, collected on the 13th day after paralysis, was negative for virus. R46064 was reported as “Sabin-like” by the intratypic differentiation tests  used in the GPLN in 2008  Table. SR46064 produced Sabin3-like results in the VDPV screening assay. R46064 gave false-negative test results because the isolate had a Sabin3 vaccine sequence in the regions corresponding to the probe and primers of the VDPV assay  Figure 2Type 2 VDPV was found in an immunocompetent girl, 3.5 years of age, in March 2014. Sabin2 was isolated from 2 stool samples collected 5 and 8 days after onset of AFP. Sabin2, isolated from stool sample 1 (R93150), was amplified in the VDPV screening assay (reported as Sabin2-like); the isolate from stool sample 2 (R93152) failed to become amplified (reported for sequencing).VP1 sequencing of R93152 showed 6 nt substitutions; therefore, it was reported as VDPV2. R93150 was also sequenced to find out whether it contained the Sabin2 homotypic mixture. VP1 sequence of R93150 showed 6 nt substitutions and no evidence of mixed bases. Substitution was not found at VP1 nt 427/428, the main target of the VDPV screening assay.Complete genome sequence analysis revealed that both isolates contained reversion of the major attenuating site at nt 481 in the 5′ UTR. The genomes of both isolates showed no recombination. Both isolates showed 15 nt substitutions in the capsid region, when compared with Sabin2  Table; oThe occurrence of VDPV2 and VDPV3 as described above may be rare. However, GPLN laboratories are unlikely to detect VDPV strains that produce false-negative results in the VDPV screening assay. False negatives are of greatest concern to the GPEI because they could impede timely detection of VDPV infections (Nucleotide substitutions in the 5′UTR and capsid region of type 2 and type 3 vaccine-derived polioviruses (VDPV), amino acid changes at neutralizing antigenic sites, and nucleotide sequence alignment of VDPV2 and VDPV3 with corresponding Sabin vaccine virus sequences.


HBG1 and HBG2 genes that are modulators of HbF synthesis, obtained in this ethnic group.Studies performed in the Andean plateau, one of the highest inhabited areas in the world, have reported that reduced availability of oxygen is associated to fetal growth retardation and lower birth weight, which are established predictors of morbidity and mortality during the first year of life. To test this hypothesis, perinatal variables of neonates born at the Juan Noé Hospital of Arica, Chile, were analyzed in relation to altitude of residence and Aymara ancestry of their mothers. The study population comprised the offspring of 5,295 mothers born between February 2004 and August 2010. Information included birth weight, height, head circumference, gestational age, altitude of residence and socioeconomic status, and was obtained from medical records. Mother´s ancestry was assessed based on surnames which were linked to percentages of Aymara admixture estimates relying on 40 selected ancestry informative markers. After correcting for the effect of multicollinearity among predictor variables, neonates born to mothers with an increased component of Aymara ancestry showed significantly higher birth weight and height at sea level, a marginally significant  decrease of birth weight and a significant decrease of height with altitude in comparison with the offspring of mothers with low Aymara ancestry. Since observed tendencies are suggestive of a possible genetic adaptation to hypoxia of the Chilean Aymara, we discuss briefly preliminary evidence related to fetal oxygen transport, particularly polymorphisms in the promoters of the  South American high altitude landscapes were colonized very recently in evolutionary terms, say during the last 400–500 generations [Human growth and development are the result of a complex interaction of genetic and environmental contributions acting during the micro-evolutionary history of human populations ,2. High Recent archaeological evidence obtained in the site of Cueva Bautista indicates that the Bolivian Highlands were colonized before 10.917 B.P.,6. The ATwo methodological shortcomings may affect altitudinal studies. First, the lack of accurate archaeological evidence concerning the time span populations have lived at high altitude may led to erroneous decisions concerning the selection of cases. Second, the failure to adequately match the ancestry of subjects residing at high and low altitudes may cause genetic differences to be wrongly attributed to altitude, when they are effectively the result of population structure.The Arica-Parinacota Region in northern Chile covers an altitudinal gradient extending from the Pacific Ocean to the Andean highlands immediately south of Peru and the west of Bolivia. Since this territory was colonized between 11,000 and 9,000 B.P. ,15,16 byWe conducted an observational, cross-sectional, descriptive and comparative study, with data collected based on non-probability sampling with specified inclusion and exclusion criteria. Information on births recorded between February 2004 and August 2010, was collected from the delivery book of Juan Noé Hospital of Arica-Parinacota Region, Arica, Chile. Socioeconomic status (SES), place of residence, maternal age, obstetric history, pregnancy-associated maternal morbidity and neonatal data including gender, gestational age, height, birth weight and head circumference of newborns were obtained from the records of the hospital. Only mothers whose deliveries were attended after 37 and before 40 weeks of gestation, whose ages were over 15 years and who gave birth to a healthy newborn with a complete set of neonatal measurements were included. A reduced group of newborns (0.43%) born to mothers with multiple pregnancies, a history of substance, cigarette and alcohol use or pathologies such as diabetes mellitus, gestational diabetes, anemia, heart disease, hypertensive disorders, intrahepatic cholestasis, mental and /or genetic problems were excluded, as well as preterm infants, microsomal and macrosomal fetuses with congenital malformations and/or associated pathology. Newborns were classified according to gender, maternal age, gestational age, residence altitude and ancestry of mothers. The following maternal age categories were considered: adolescents (under 20 years), youth (20 to 27 years), young adults (28 to 39 years) and adults (≥ 40 years). The National Health Fund (FONASA), which classifies individuals into four groups according to household income weighted by number of family members, provided socioeconomic statuses of participating mothers. Place of residence was obtained from medical record and classified according to altitude into two groups (1) coast and lower mountain region  designated as lowland and (2) Andean plateau (3000–4300 m) designated as highland. These two geo-ecological zones differ in population density, atmospheric pressure, diurnal temperature variation, humidity, solar and cosmic radiation . Aymará Three morphological variables  were investigated in this study: birth weight (in kg), height (in cms) and head circumference (in cms). In addition we recorded gestational age (in weeks).Hypotheses testing were conducted with ANOVA, Student’s t-tests for independent samples and multiple regression analysis. Data analyses were performed using SPSS routines of descriptive statistics.This research was approved by the Scientific Ethics Committee of the Chilean Ministry of Health on July 20, 2007 and ratified by the Bioethics Committee of the Health Service of the Region of Arica and Parinacota and the Dr. Juan Noé Crevani Hospital Bioethics Committee. The patients records/information was anonymized and deidentified prior to analysis.The total sample comprised 5,295 newborns . 2,621 .The degree of association among the three neonatal morphological measurements was examined using pairwise (Pearson) correlations . As expeThe effects of altitude of residence (lowland versus highland) and ancestry (LAGB versus HAGB mothers) on the three examined morphological variables revealed that children born to HAGB mothers were in general heavier, taller and had larger head circumferences than children born to LAGB mothers . GestatiIn order to correct for the possible effect of multicollinearity among predictor variables we conducted multiple regression analyses .Summarizing results, we noted that altitude showed a significant negative effect on birth weight  and on height  independently of gestational age, maternal age, gender and Aymara ancestry. Conversely, if altitude was held constant together with the remaining predictors, Aymara ancestry showed a marginally significant positive effect on birth weight  and a significant positive association with birth height .These results confirm the existence of an altitudinal effect on birth weight and height and of Aymara ancestry on height. Altitude and ancestry showed no effects on head circumference.of the same ethnic group living at different altitudinal levels was compared, allowing to simultaneously examine the effects of different altitudinal levels and contrasting percentages of ancestry. Furthermore, we note that the Aymara have been living at high altitude for more than 350 generations, enabling genetic variants that were either slightly advantageous or neutral at low altitude, to turn advantageous and increase their frequency. A limitation of this study is that in spite of the availability of medical records from 5,295 newborns and almost no missing data, the number of offspring born to highland residence mothers (n = 54) was small as a consequence of a depopulation of this eco-geographic area due to massive migration to the city of Arica in search of job opportunities. This caused lack of power and forced us to pool mothers with one and two Aymara surnames, a procedure that may have underestimated the true effects of Aymara ancestry on neonatal variables.An important strength of this study is that the status of neonates born to mothers The identified effects of Aymara ancestry have implications in relation to adaptation to hypoxia. One feature consistently related to altitude adaptation is the small body size and lower body weight of newborns at high altitude ,19. Our VEGF and ELDT 1B, which was linked to cardiophysiopathological findings. Unfortunately the possibility that differences observed between Aymara and Wichi are due to population structure have not be ruled out. Consequently further research is necessary.Recently, several studies have reported the involvement of genes or genomic regions conferring adaptation to hypoxia. For example, exome genome sequencing of Tibetan and Andean highlanders suggested a role in adaptation to hypoxia of the endothelial Per-Arnt-Sim (PAS) domain protein 1 (EPAS1) gene, which acts as a transcription factor involved in response to oxygen deficiency ,33,34,35Returning to our results, we note that besides supporting previous findings based on interethnic highland birth weight comparisons, the design of this study allowed us to apportion environmental  and genetic (Aymara) effects on neonatal variables. For example, weight differences by altitude (low versus highland) amounted 3.49–3.30kg = 190 grams  among newborns to HAGB mothers, and 180 grams (non-significant) among newborns to LAGB mothers. Weight differences by Aymara ancestry (HAGB versus LAGB) amounted 50 grams  at lowland, and 40 grams (non-significant) at highland. The contribution of additional, environmental factors such as nutrition and culture to these differences is probably rather small, since LAGB and HAGB mothers living at sea level did not belong to significantly different SES categories and live, after undergoing a substantial process of transculturation, in the westernized city of Arica.HBG1 and HBG2 genes are modulators of HbF synthesis, as a 4-basepair deletion of AGCA from nucleotide positions −222 to −225 in the promoter of the HBG1 gene, which is associated with diminished expression of fetal γ globin chains and might therefore affect fetal oxygen supply at high altitude [HBG2−158 T allele, are associated with a delayed fetal to adult globin switch after birth [HBG1 and HBG2 genes were sequenced, the 4-basepair AGCA deletion of HBG1 promoter was less frequent in the Aymará than in lowlanders . The T allele of the −158 C > T polymorphism in the HBG2 promoter was also decreased in Aymará, compared with lowlanders . A combined analysis of both markers showed that none of the Aymará with the AGCA deletion carried the HBG2 T allele, in contrast to 43% of lowlanders (P = 0.030). Although still preliminary, reported results are interesting and warrant more thorough investigations.Some polymorphisms in the promoters of the altitude ,39. Elevaltitude . Both thaltitude  and the er birth . In ordeer birth  comparedWhen pregnancy occurs at high altitude, probably hypoxia decreases birth weight and increases morbidity and mortality during the first year of life. However, a higher percentage of Aymara admixture confers some protection to newborns, reducing the influence of high altitude through an increase in height and birth weight. Since this effect is also expressed in newborns of Aymara mothers living at sea level, the participation of genetic or epigenetic factors is suggested. Particularly interesting are recent findings linking maternal PRKAA1 and EDNRA genotypes to birth weight, uterine artery diameter and metabolic homeostasis at high altitude  reinforc


The cingulum and fornix play an important role in memory, attention, spatial orientation, and feeling functions. Both microstructure and length of these limbic tracts can be affected by mental disorders such as Alzheimer’s disease, depression, autism, anxiety, and schizophrenia. To date, there has been little systematic characterization of their microstructure, length, and functional connectivity in normally developing brains. In this study, diffusion tensor imaging (DTI) and resting state functional MRI (rs-fMRI) data from 65 normally developing right-handed subjects from birth to young adulthood was acquired. After cingulate gyrus part of the cingulum (cgc), hippocampal part of the cingulum (cgh) and fornix (fx) were traced with DTI tractography, absolute and normalized tract lengths and DTI-derived metrics including fractional anisotropy, mean, axial, and radial diffusivity were measured for traced limbic tracts. Free water elimination (FWE) algorithm was adopted to improve accuracy of the measurements of DTI-derived metrics. The role of these limbic tracts in the functional network at birth and adulthood was explored. We found a logarithmic age-dependent trajectory for FWE-corrected DTI metric changes with fast increase of microstructural integrity from birth to 2 years old followed by a slow increase to 25 years old. Normalized tract length of cgc increases with age, while no significant relationship with age was found for normalized tract lengths of cgh and fx. Stronger microstructural integrity on the left side compared to that of the right side was found. With integrated DTI and rs-fMRI, the key connectional role of cgc and cgh in the default mode network was confirmed as early as birth. Systematic characterization of length and DTI metrics after FWE correction of limbic tracts offers insight into their morphological and microstructural developmental trajectories. These trajectories may serve as a normal reference for pediatric patients with mental disorders. The limbic system is a group of interconnected cortical and subcortical structures dedicated to linking visceral states and emotion to cognition and behavior Mesulam, . The limIn the developing human brain, water molecules tend to diffuse more freely along the limbic tract, instead of perpendicular to it. This diffusion property can be measured non-invasively with diffusion MRI, a modality of MRI. Diffusion tensor imaging (DTI)   from 15 of the subjects were acquired. After cgc, cgh, and fx were traced with DTI tractography, absolute and normalized tract lengths and DTI-derived metrics including FA, MD, AD, and RD were measured for traced limbic tracts. Free water elimination (FWE) algorithm  with age range from 0 month (birth) to 25 years participated in this study. The age range covered the important developmental periods of early childhood, childhood, adolescence, and young adulthood. Figure  M/24 F wb-value = 1000 s/mm2, and repetition = 2. The axial image dimension was 256 × 256 after reconstruction. The total acquisition time of DTI was 11 min. With 30 diffusion weighted image (DWI) volumes and 2 repetitions, we accepted only those scanned datasets with less than 5 DWI volumes affected by motion more commonly seen in scan of neonates and young children. The affected volumes were replaced by the good volumes of another DTI repetition during postprocessing. rs-fMRI was also acquired from 15 out of 65 subjects, including 3 neonates and 3 young adults. A T2* weighted gradient echo EPI sequence was used. A total of 210 T2* weighted whole brain volumes were acquired with the following parameters: TR = 1500 ms for neonates and 2000 ms for young adults, TE = 27 ms for neonates and 30 ms for young adults, flip angle = 80°, in-plane imaging resolution = 2.4 mm × 2.4 mm for neonates and 3.4 mm × 3.4 mm for young adults, and slice thickness = 3mm with no gap for neonates and 4 mm with no gap for young adults; slice number = 30 for neonates and 37 for young adults. Scan time of rs-fMRI was 5–6 min. Rs-fMRI and DTI images were acquired in the same session.MRI of 3 neonates (2 M/1 F) at 40 weeks of gestation and 10 adolescents  was acquired from Children’s Medical Center at Dallas. MRI of all other subjects was acquired from Children’s Hospital in Beijing. A Philips 3 T Achieva MR scanner at both sites was used. A single-shot echo planar imaging (EPI) with SENSE  parallel imaging scheme was adopted for DTI acquisition. Eight-channel SENSE head coil and consoles installed with R2.6.3 software were used for both sites. The DTI imaging parameters were: TR (repetition time) = 6850 ms, TE (echo time) = 78 ms, in-plane field of view = 168 mm × 168 mm for neonates and 224 mm × 224 mm for all others, in-plane imaging resolution = 1.5 mm × 1.5 mm for neonates and 2 mm × 2 mm for all others, slice thickness = 1.5 mm for neonate to 2 mm for all others, slice number = 60–65 depending on the height of the subject brains, 30 independent diffusion-weighted directions  linear registration with automated image registration (AIR) algorithm  .The absolute and normalized limbic tract length were measured based on the tractography results. Specifically, geodesic length of each fiber in cgc, cgh, and fx was measured. A histogram was established for each limbic tract to show the distribution of fiber length. Note that a tract contains fibers with different lengths. In order to reduce the bias caused by a large number of short fibers, we defined the tract length by averaging 10% longest fibers of each tract. As shorter fibers could only cover part of the tract, reconstruction of the top 10% longest fibers showed that morphology of these fibers was in good agreement with that of the entire tract of all fibers. Tract length was further normalized by the size of the brain represented by the length between most anterior to most posterior edge to quantify relative tract length change during development.S is the measured signal in the voxel, fiso is the volume fraction of the isotropic compartment, b is the applied b-value, S0 is the signal with no diffusion weighting, Diso is the diffusion coefficient of the isotropic compartment, and fiso by solving the non-linear equation in the model. The initial values of Diso was 3 × 10−3 mm2/s  of the rs-fMRI data. ICA was conducted with MELODIC tool y  and age t, y = f(t) + c·sex + ε, where f(t) was linear, logarithmic, or polynomial function of t and ε was an error term. F-test was used to find the best fitting curve. We first tested if the measurement y was age dependent, if so, we further tested which of the following curves, linear, logarithmic, or polynomial curves, fitted the data best.The following equation was used for fitting a model between measurement With age and gender as covariates, generalized linear model (GLM) was used to test (1) if FWE correction significantly changed DTI metric measurements and (2) if lateralization  was a significant factor for all measurements. The null hypothesis for FWE correction is that there is no difference of DTI metric measurement before and after the correction. The null hypothesis for lateralization is that there is no difference of length or DTI metric measurement of corresponding limbic tracts in the left and right hemisphere. The Bonferroni correction was used to control the spurious positives when rejecting the null hypotheses.As shown in Figure R2 was obtained with the polynomial fitting.The plots of uncorrected and corrected DTI metrics, namely, FA, MD, AD, and RD, of all limbic tracts at different ages are shown in Figures As shown in Table R2. After F-test, significant difference between logarithmic and linear model was found only for absolute length of cgc-L  signal oscillations are temporally correlated can be either directly connected by white matter tracts or connected through a relay. The adult brain regions of PCC, MPFC, and MTL in DMN are connected by limbic tracts (Greicius et al., It is difficult to acquire all DTI data from birth to young adulthood in one site. We combined datasets acquired in two sites, but with the same imaging protocol and same Philips 3 T Achieva scanners. Rigorous quality control is conducted routinely for both scanners. We performed quantitative analysis based on DTI tractography and metric measurements to evaluate the effects of these two scanners in our previous study (Huang et al., In conclusion, with DTI and rs-fMRI from developing brains at cross-sectional ages from 0 month to 25 years, we have quantitatively characterized development of microstructure, length, and connection of limbic tracts that play a key role in emotion, memories, and behavior. Logarithmic developmental trajectories were found for most age-dependent changes of limbic tract microstructures and lengths from birth to young adulthood. Correction of CSF contamination has significantly increased FA and reduced MD, AD, and RD of developing limbic tracts, but has not changed the age-dependent trajectories of these metrics. Stronger microstructural integrity and longer tract lengths were found for limbic tracts on the left side compared to those on the right side. Consistent role of limbic tracts connecting major DMN regions at birth and adulthood has been confirmed and demonstrated.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.http://www.frontiersin.org/Journal/10.3389/fnagi.2014.00228/abstractThe Supplementary Material for this article can be found online at Click here for additional data file.


Esophageal bronchogenic cysts are extremely rare. Here we report a more rare type of both presence of intra- and paraesophageal bronchogenic cyst that was safely removed via surgical resection. A 31-year-old male patient with space-occupying lesions in the mediastinum suddenly presented with persistent chest pain for 2 days and then transferred to dysphagia >1 week. Preoperative diagnosis is difficult. Endoscopic ultrasonography (EUS) showed a hypoechoic cystic-solid mass arising from the muscularis propria and local hyperechoic area in the deeper portion of cyst, concomitant with a heterogeneous center and tube-like structure lesion in mediastinum. Turbid coffee color paste contents were aspirated inside the tumor under endoscopic ultrasonography guided-fine needle aspiration (EUS-FNA). A subsequent surgery was performed and histologic finding was diagnostic of esophageal bronchogenic cyst. Immunohistochemical staining confirmed the cyst was positive for carbohydrate antigen 199 (CA199) and carbohydrate antigen 125 (CA125). At a follow-up visit 3 months later, the patient had a regular diet and no complaint. This study is to summarize the clinical manifestations and EUS features of esophageal bronchogenic cyst by retrospectively reviewing the literature and simultaneously to provide guide for the correct examination scheme.The appearance of esophageal bronchogenic cyst can be great variation; EUS seems to be a valuable option for diagnosis and surveillance. Sporadic cases concerning intramural esophageal bronchogenic cysts are reported.5 Moreover, despite improved imaging modalities, a definitive preoperative diagnosis is difficult. To our knowledge, this is the first report to share our experience with a very rarely clinical entity of both periesophageal and intramural esophageal bronchogenic cysts with positive expressions of CA199 and CA125 by immunohistochemical study and also review of the literature with an emphasis on endoscopic ultrasonography image characteristics to guide correct examination scheme. (The answer to Reviewer #1[1] and [2].) Patients have signed informed consent and data had been anonymized and deidentified.Bronchogenic cysts are rare congenital malformation originated from the development of ventral foregut, and mostly located in the middle and superior mediastinum.A 31-year-old man who presented to our hospital with suddenly symptoms of persistent chest pain for 2 days and then transferred to dysphagia >1 week. He has no fever, cough, vomit, and weight loss. And there was no obvious positive characteristics on physical exams. With regard to laboratory values, only serum tumor marker CA125  and CA199  levels were remarkable increased. Blood routine test, erythrocyte sedimentation rate (ESR), C-reactive protein, hepatic and renal function were all within normal limits. An esophagography obtained before presentation identified a huge filling defect compressed to esophagus under the aortic arch . Here we reported a more rare type of both presence of intra- and para-esophageal bronchogenic cyst that was safely removed via surgical resection. To our knowledge, a similar case had never been reported.Bronchogenic cysts are broncho-pulmonary foregut abnormalities that usually occur along the tracheobronchial tree in the mediastinum. They may be located in paratracheal, carinal, or hilar.10 Despite improved imaging modalities, a definitive diagnosis remains difficult preoperatively. CT or MRI sometimes can indicate the nature of the cyst. But once bronchogenic cyst is infected or high in protein or calcium content, its density may fall into the same solid mass scope that increases diagnostic uncertainty.12 Moreover, the intramural and extramural relationship of esophageal cysts is also difficult to identification. EUS is sensitive for distinguishing cystic from solid masses and facilitates the diagnosis of intramural esophageal lesions because of its capacity to clearly delineate the size, layer of origin, and the relationship of the cyst to the esophagus.3 The present study first to review of the literature with an emphasis on EUS appearance of esophageal bronchogenic cyst, as only 8 cases have been presented in the literature since 2000s17  is a newly effective and safe procedure to treat submucosal tumors originating from the muscularis propria and has been successful to extract an intraesophageal bronchogenic cyst .23 This treatment would be a less complicated and less risky choice, but long-term follow-up visits and complications are required to evaluated further, EUS seems to be a valuable option for diagnosis and surveillance.With regarding to the management of esophageal bronchogenic cysts, complete surgical removal by thoracotomy or video-assisted thoracoscopy is recommended, even when they are asymptomatic, because of subsequent complication of infection, rupture, intracystic hemorrhage, carcinomatous change.Chaoqun Han: collection and assembly of data, data analysis, manuscript writing, performed the literature search; Rong Lin: collection and assembly of data, manuscript writing and performed the literature search; Jun Yu: provision of pathological data and valuable suggestions; Qin Zhang: provision of pathological data and valuable suggestions; Yang Zhang: provision of gross samples post operation; Jun Liu: administrative support and valuable suggestions; Zhen Ding: conception and design, financial support, revised the manuscript as corresponding author; and Xiaohua Hou: financial support and valuable suggestions.


Am J Trop Med Hyg 92:1038–1044 by Schoeps and others, there is a small layout error in Table 2, which may cause misinterpretation. The variable “year of birth” is indented, so it appears to be one category of the above variable “twin birth”. “Year of birth” is a separate variable that is not part of “twin birth”. The journal regrets the error.In


Identifying and understanding the current sources that give rise to bioelectric fields is a fundamental problem in the biological sciences. It is very difficult, for example, to attribute the time-varying features of an electroencephalogram recorded from the head surface to the neural activity of specific brain areas; model systems can provide important insight into such problems. Some species of fish actively generate an oscillating (c. 1000 Hz) quasi-dipole electric field to communicate and sense their environment in the dark. A specialized electric organ comprises neuron-like cells whose collective signal underlies this electric field. As a step towards understanding the detailed biophysics of signal generation in these fish, we use an anatomically-detailed finite-element modelling approach to reverse-engineer the electric organ signal over one oscillation cycle. We find that the spatiotemporal profile of current along the electric organ constitutes a travelling wave that is well-described by two spatial Fourier components varying in time. The conduction velocity of this wave is faster than action potential conduction in any known neuronal axon (>200 m/s), suggesting that the spatiotemporal features of high-frequency electric organ discharges are not constrained by the conduction velocities of spinal neuron pathways. The production and detection of electric fields is essential for living organisms from the single-cell to humans12347810et al.The EODs of weakly electric fish come in two forms9121312151819201111The electric organ (EO) comprises large populations of electrocytes that generate the current underlying the EOD. Electrocyte characteristics vary across species122526A. leptorhynchus, a high-frequency (c. 1000 Hz) wave-type fish, and (2) determine the current sources produced by the electric organ. Previous studies1128To better understand the biophysical mechanisms and constraints underlying high-frequency neurogenic EODs, we use a finite-element modelling (FEM) approachA. leptorhynchus is complex relative to other weakly electric fish1810The spatiotemporal electric field of To make accurate predictions for electric sensing (electric images)13203031102) so the y-axis is clipped to 50 mA/m2 in 2  that minimized the RMSE between data and model for each of 25 EOD phases (see Methods). Fitting the model in this way provides a prediction for the spatiotemporal variation of current density on the EO. /m2 in 2  reveals nd and 3rd components (1st and 2nd harmonics) increase significantly. This is a reflection of the frequency doubling observed previouslyGymnotus carapo, the electric organ comprises separate functional segments, each producing distinct waveforms1124A. leptorhynchus, but the data shown here are more consistent with EO activation resembling a continuous travelling wave.We first consider the temporal variations in current density at distinct locations along the EO. To further explore the travelling wave hypothesis, we next performed a Fourier analysis in space . This involves considering a single point in time (single EOD phase) and then characterizing the spatial current profile over the EO by the amplitudes and phases of different spatial Fourier components. Repeating for all EOD phases provides a description of the evolution of EO activation. For reference, a pure oscillating dipole could be explained by a single spatial frequency component with a constant phase, but with amplitude varying in time. On the other hand, a phase that changes continuously in time is suggestive of a travelling wave.st and 2nd harmonics). For clarification, the 1st harmonic here is defined as the Fourier component with wavelength equal to the EO length. We found that the primary features of the EO current density were well-represented (~3.8% RMSE) by only two spatial Fourier components . Increased resistance could arise from a high resistance membrane  used in our fits could over-fit the data, the resulting continuity of current density between phases implies that this is not the case. Overall, this suggests that the solutions, and thus our predictions, are robust.BrachyhypopomusA. leptorhynchus.The electric fields generated by weakly electric fish are often compared to those resulting from an oscillating dipole for which the polarity of the head and tail are of opposite sign. However, in many species, this analogy cannot explain the observed spatiotemporal complexitiesOne shortcoming of our model is that it considers only a 2D cross-section of the 3D electric field. It is thus possible that accounting for current flow outside of the horizontal plane will allow for a more parsimonious EO current density. However, the width (right-left) of these fish is small compared to the height  making the lateral skin surface relatively flat; as a result, the current flow is predominantly in the horizontal plane, so our 2D approximation may not be a major problem. A simpler 2D model has been used to accurately locate weakly electric fish in shallow water using recordings from an electrode array1036Weakly electric fish have been an effective model system in which to explore various aspects of bioelectric signalling. Most prominent is the work on electric field perturbations that provide a basis for electric imaging. The electric images produced by simple objects, investigated using both experimental and computational approaches, have provided novel insights into how electric fields can be used for object localization and identification134445The problem of bioelectric source identification has received less attention in the electric fish literature24Apteronotus leptorhynchus (brown ghost knife fish) is based on previous work focused on a single phase of the EOD cycle18101111Gnathonemus petersiiet al.The finite-element approach presented here to model the electric potential of In order to calculate the electric potential (and electric field), Poisson’s equation was solved in three dimensions using COMSOL Multiphysics v5.0 :J is the total current flux through an element . In order to restrict the model to two dimensions, the conductivity along the dorsoventral axis was set to 0 S/m. For the low frequencies used by the fish (<10 kHz), capacitive effects are negligible and the fish body is mostly resistive8where The electric organ (EO), modelled as a caricature of the organ described by Bennett1118i is the amplitude, σi is the standard deviation and xi is the mean of the ith Gaussian . Babineau et al.where αThe experimental data from Assad and Rasnow quantify the electric field and potential over 25 phases of the EOD cycle . At each phase, measurements were made with varying spatial resolution on one side of the fish from 8 cm rostral to the head, to 5.2 cm caudal to the tail and up to 14 cm lateral to the fish :where where I) across all EO boundaries over the entire EOD cycle. For the solutions presented in this paper, net current was −0.36 A/m2, less than 0.1% of the total current flux Current conservation was evaluated by calculating the net current flow .


Preterm birth is associated with increased risk of asthma-like symptoms and purchase of prescription asthma medication in childhood. We investigated whether this association persists into adulthood and whether it is affected by accounting for neonatal respiratory morbidity (acute respiratory disease and bronchopulmonary dysplasia).A national cohort of all infants born in Denmark in the period 1980–2009 was included in this register study. Data on purchase of asthma medication  in 2010–2011 were obtained from the Danish National Prescription Registry. Associations between gestational age (GA), neonatal respiratory morbidity and a cross-sectional evaluation of asthma medication purchase were explored by multivariate logistic regressions.A full dataset was obtained on 1,790,241 individuals, 84.6% of all infants born in the period. Odds-ratios (95% CI) for the association between GA and purchase of asthma medication during infancy were: 3.86 (2.46–6.04) in GA 23–27 weeks, 2.37 (1.84–3.04) in GA 28–31 weeks and 1.59 (1.43–1.77) in GA 32–36 weeks compared to term infants with GA 37–42 weeks. Associations weakened in older age groups and became insignificant in young adults born extremely and very preterm with odds-ratios: 1.41 (0.63–3.19) and 1.15 (0.83–1.60) in GA 23–27 and 28–31 respectively. When adjusting for neonatal respiratory morbidity, the associations weakened but persisted both in childhood and adolescence.There was a strong dose-response association between gestational age and the purchase of prescription asthma medication in infancy and childhood. This association weakened during adolescence and was mostly non-significant in young adulthood. The increased risk of prescription asthma medication purchase in ex-preterm children could only partly be explained by neonatal respiratory morbidity. Prematurity  <37 weeks) is associated with a high prevalence of neonatal respiratory distress compared to term birth. Nevertheless, most of these infants only require a few days or weeks of respiratory support and do not necessarily have any significant respiratory symptoms when they reach an age equivalent to term. However, the extremely preterm infants with GA 23–27 weeks are significantly more vulnerable. Their oxygen dependency tends to persist for many weeks and most of them develop some degree of bronchopulmonary dysplasia (BPD) . BPD is Even though the vast majority of all preterm infants do not develop BPD, it has been shown that preterm birth is a risk factor for non-specific asthma-like symptoms in childhood –8. HowevWe therefore aimed to reassess the association between the full range of prematurity and the purchase of prescription asthma medication as a marker for the prevalence of a persistent disease burden of asthma-like symptoms across the age span from infancy to young adulthood. We included an assessment of the influence of neonatal respiratory morbidity on this association, in order to evaluate the potential link between neonatal respiratory morbidity and later respiratory symptoms. For this purpose, we conducted a register based study linking register data on an individual level in a Danish national cohort.st 2010 to December 31st 2011, when subjects were 0–31 years of age. The population in Denmark consists of 5.6 million inhabitants. A unique Central Personal Registration (CPR) number is assigned to all individuals at birth, to which all register entries on social and health services are consistently and unambiguously linked through the Civil Registration System [We conducted an observational retrospective register-based study of a national cohort of all infants born in Denmark in 1980–2009 with a cross-sectional follow-up in the period January 1Data were retrieved from Statistics Denmark in April 2012. The study population was divided into six separate age groups defined by age in years at follow-up (birth years): infants 0–2 (2009), preschool children 3–5 (2006–2008), children 6–11 (2000–2005), adolescents 12–17 (1994–1999), adults 18–24 (1987–1993) and 25–31 (1980–1986).Data on all purchases of prescription medication from all non-hospital pharmacies during the two-year follow-up period 2010–2011 were obtained from the Danish National Prescription Registry. The register was established in 1994 and data are available through Statistics Denmark from 1995 . Hospita- either a combination of at least one purchase of inhaled selective β-2 receptor agonist (R03AC) AND at least two purchases of one of the following other drugs for obstructive airway disease: inhaled glucocorticoids (R03BA), inhaled anticholinergics (R03BB), theophyllines (R03DA), oral leukotriene-receptor antagonists (R03DC), systemic steroid (H02AB), a combination inhaler (R03AK)—or at least two purchases of a combination inhaler containing a long-acting β-2 receptor agonist and either glucocorticoids or anticholinergics (R03AK).Our two exposures of interest were GA and neonatal respiratory morbidity.GA was retrieved from the Medical Birth Registry. The variable was evaluated in full weeks and is reported to the register as assessed by a midwife and/or doctor. The register was established in 1973. In 1978 mid-wives’ reporting to the register was made mandatory by law. Before 1997 the reporting was made directly to the Danish Health and Medicines Authority. From 1997 onwards, the register has been based on a combination of a subset of the National Patient Register , the CivNeonatal respiratory diagnoses were obtained from the National Patient Register during the half-year following the date of birth. The register was established in 1977 and contains information about patients from all public Danish hospitals. Data from an emerging private sector were included in the register from 2000 onwards. There are no private maternity or neonatal units in Denmark. Additionally, the register became the basis for payment of both public and private hospitals through the Diagnostic Related Group (DRG) system in 2000 .The International Classification of Diseases ICD-10 system replaced the ICD-8 system in the register in 1994. Neonatal respiratory diagnoses were split into two groups. One group defined as “acute neonatal respiratory distress” included the diagnoses (ICD codes) of respiratory distress and transitory tachypnea of the neonate (ICD-8 776.290 / ICD-10 P22.1 and P22.8), respiratory distress syndrome (ICD-8 776.190 / ICD-10 P22.0) and neonatal emphysema (ICD-8 776.200 / ICD-10 P25.0). The second group was defined as “BPD” or chronic neonatal respiratory disease. It included the diagnoses of pulmonary fibrosis (ICD-8 517.010) and bronchopulmonary dysplasia (ICD-10 P27.1). The ICD-8 diagnosis of pulmonary fibrosis given in the neonatal period is assumed to be equivalent to the ICD-10 diagnosis BPD. In the following, the term BPD will be referring to both the ICD-8 and ICD-10 diagnoses.Well-known risk factors for persistent respiratory symptoms, which were available in the registers, were also included in the multivariate analyses as potential confounding or mediating variables as follows.The Medical Birth Registry provided data on the following birth-related exposure variables in addition to GA:birth weight (BW), from 1979 in 10-grams groups and from 1991 in four digits. Considering the collinearity between GA and BW, the variable small-for-gestational age (SGA) was used in our analyses instead of BW. SGA was defined as a BW lower than -2SD from the reference value sexolder siblingsmultiple birthdelivery method, i.e. cesarean section (subdivided into emergency and elective sections from 1982 onwards) or natural birthmaternal smoking during pregnancy .Data on maternal educational level at the time of birth were obtained from the Register of Education of the Population. The data are provided through Statistics Denmark from 1981 in collaboration with the Danish Ministry of Education . They argroup 1 included ISCED levels 0–2: early childhood educational development, pre-primary education, primary education and lower secondary education.group 2 included ISCED levels 3–4: upper secondary education and post-secondary non-tertiary education.group 3 included ISCED levels 5–8: short-cycle tertiary education, Bachelor’s or equivalent level, Master’s or equivalent level, Doctor or equivalent level.For children born in 1980, maternal education level was retrieved from 1981 data. When data on maternal education were missing, data on paternal education were used instead.Maternal asthma medication was defined as described for the cases. Maternal atopic medication was defined as at least one purchase of nasal steroid spray (R01AD), facial steroid cream group I (D07AA) or steroid cream group III (D07AC), all of which required a prescription. Since no medication data were available before 1995, maternal medication purchase was evaluated during the follow-up period in 2010–2011. Thus, it was not an indicator of maternal medication use during pregnancy but of asthmatic and atopic hereditary predisposition.Deaths were obtained from the Causes of Death Register, which contains detailed registration of all deaths in Denmark. The registration of Death Certificates was computerized in 1970 and since 2007 the certificates have been systematically issued and reported electronically. The register is matched with data from the Civil Registration System to ensure its validity .Logistic regression analyses were performed in four setups with purchase of prescription asthma medication as the outcome variable:Unadjusted.Model 1: Adjusted for birth-related and socio-economic variables: sex, SGA, older siblings, multiple birth, maternal educational level, caesarean section, maternal asthma medication, maternal atopy medication and maternal smoking.Model 2: As model 1 and in addition including acute neonatal respiratory distress.Model 3: As model 2 and including BPD, i.e. all neonatal respiratory morbidities.Only subjects with complete data were included in these analyses. SAS 9.3, SAS Institute was used for all statistical analyses.Data extraction for the present study has been approved by the Danish Data Protection Agency and the Danish Health and Medicines Authority (Jr.no 2007–41–0806). Data were processed in accordance with Denmark’s Act on Processing of Personal Data, which allows linkages in this type of study without individual ethical approval.The anonymized data are available from Statistics Denmark. Access can be granted to Danish research environments that are pre-approved by Statistics Denmark or to foreign researchers with an affiliation to a Danish authorized environment. Whilst we cannot guarantee such an access, we can assist interested institutions in any request in this context.st 1980 and December 31st 2009. The missing 326,831 (15.4%) individuals were excluded due to stillbirth, missing data on either GA, BW or both GA and BW, BW-values differing +/- more than 3 standard deviations from reference values published by Marsál et al. [Our study population of 1790241 individuals constituted 84.6% of the total of 2,117,072 individuals born in Denmark between January 1 326,831 .4% indivIn the study population, 96,778 (5.4%) of the individuals were born preterm: 2,247 extremely preterm (GA 23–27), 9,626 very preterm (GA 28–31) and 84,905 moderately preterm (GA 32–36). The sex distribution was 917,010 males (51.2%) to 873,321 females (48.8%). The number of individuals included in each age group is indicated in and at least two other anti-asthmatic drugs or at least two purchases of combination inhalers) is available in A description of the proportions of each medication category purchased by individuals meeting our outcome definition criteria  compared to 7.0 (CI 6.7–7.3) in girls at 0–2 years of age), and higher prevalences in women in adulthood (2.4% (CI 2.3–2.5) compared to 1.8% (CI 1.8–1.9) in men at 25–31 years of age) .BPD was predominantly present in the extremely preterm and very preterm. The extremely preterm demonstrated high BPD prevalences, ranging from 34.7% (CI 30.7–38.8) to 43.6% (CI 39.9–47.3) in the young age groups. In the adult age groups, BPD prevalence in the extremely preterm was considerably lower: 4.8% (CI 2.9–6.9) in 18–24 years and 8.3% (CI 4.5–12.1) 25–31 years. Acute neonatal respiratory disease was present in all preterm groups, but was overrepresented in the extremely preterm with prevalences from 72.3% (CI 68.1–76.6) in 18–24 years to 89.0% (CI 85.5–92.5) in 3–5 years . Acute nUnadjusted ORs and 95% CIs (in parenthesis in the following) were above 1 for the preterm groups during childhood (age groups <18 years) with a dose-response relationship to GA : They raMost covariates had stable associations with the outcome throughout all analyses . Thus, nThe results from the four different model setups are illustrated in In this large long-term follow-up study, we found a strong association between gestational age at birth and purchase of prescription asthma medication in childhood. The association increased in a dose-response manner with decreasing gestational age, but weakened in higher age groups and was small or non-existent in adult groups.In the moderately preterm group a weak association of GA with asthma medication persisted with ORs from 1.59 in infancy to 1.14 in adulthood. The fifty percent increased risk of purchasing prescription asthma medication in preschool children born moderately preterm is an important finding which also supports data from another recent study . Our finIn the extremely preterm and very preterm groups, the associations of GA to asthma medication were strong in preschool-aged children. These associations weakened gradually in our school age and adolescent groups. In adulthood, ORs but not CIs remained above 1. Our findings corroborate the results of previous studies showing increased risk of airway disease in childhood in these individuals . Our stuThe choice of asthma medication varied in the GA-groups . IndividWe adjusted our multivariate analyses for neonatal respiratory morbidities, which we considered possible mediators, but these could not account for the full effect of prematurity. The association of prematurity per se with purchase of prescription asthma medication in our preschool- and school-aged children suggests that other factors associated with extreme and very preterm birth must also play an important role. Possible factors include disturbances in lung growth and alveolarization either due to the processes—including maternal risk factors–that lead to preterm birth and low birth weight or due to premature exposure to extrauterine conditions –25. FurtIn the adult groups, the increased risk of asthma medication purchase in individuals born extremely and very preterm was not statistically significant. However, we believe it is important to bear in mind that extreme and very preterm birth is associated with increased airway responsiveness and some degree of pulmonary impairment in young adulthood . It has Cesarean section was included as a possible exposure, confounder or mediator in our multivariate analyses. Previous studies have shown that emergency sections have a stronger association with purchase asthma medication than elective sections ,34. Our The interpretation of our findings is limited by the fact that the definition of BPD and the severity of the histopathological changes associated with BPD have changed over time ,36. A chAnother register limitation is the lack of data on maternal smoking before 1991. However, the variable did not show significant associations with asthma medication from 6–17 years. Furthermore, associations between GA and purchase of asthma medication were not significant in the oldest age groups, with the exception of the moderately preterm. Finally, missing values don’t exceed 5% for this variable in each age group from 1991 onwards . TherefoSome further limitations associated with our study design should be considered. First, the study is a cross-sectional evaluation of asthma medication purchase in different age groups. Thus, the decrease in prevalence in the older age groups may be a cohort effect: the younger age groups may have been more exposed to the real etiological factors that disturb lung growth and development compared to the older age groups. Prematurity per se, as well as both neonatal acute respiratory morbidity and BPD are possibly mere indicators of risk. Second, the study being register based, maternal prescription atopic medication was used as a proxy for atopic hereditary predisposition, which introduced a risk of misclassification of this variable. However, we found a prevalence of 17.9% in the maternal population comparable to prevalences of atopic eczema (17.7–22.4%)  and atopThe strengths of this study are its large size and the high reliability of data from central databases individually linked by a unique CPR number, allowing investigation of the full spectrum of prematurity and several important confounders and mediators over a significant period of time. Furthermore, the register provenance of the analyzed data allows a minimization of selection and recall biases. Last, we believe that prescription of asthma medication in Denmark follows frequently updated national guidelines ,44 with We conclude that there was an association between gestational age and the purchase of prescription asthma medication in infancy and childhood, which appeared to weaken during adolescence and early adulthood. This association was only partly explained by neonatal respiratory morbidity diagnoses, indicating that other causal pathways exist. It remains to be investigated whether this weakening association occurs in longitudinal studies, and whether the strength of this association will increase again later in adulthood.S1 TableGA: Gestational Age. ATC: Anatomical Therapeutic Chemical (classification system).(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.


Few studies on cluster-based synthetic effects of multiple risk factors for birth defects have been reported. The present study aimed to identify maternal exposure clusters, explore the association between clusters of risk factors and birth defects, and further screen women with high risk for birth defects among expectant mothers.Data were drawn from a large-scale, retrospective epidemiological survey of birth defects from 2006 to 2008 in six counties of Shanxi Province, China, using a three-level stratified random cluster sampling technique. Overall risk factors were extracted using eight synthetic variables summed and examined as a total risk factor score: maternal delivery age, genetic factors, medical history, nutrition and folic acid deficiency, maternal illness in pregnancy, drug use in pregnancy, environmental risk factors in pregnancy, and unhealthy maternal lifestyle in pregnancy. Latent class cluster analysis was used to identify maternal exposure clusters based on these synthetic variables. Adjusted odds ratios (AOR) were used to explore associations between clusters and birth defects, after adjusting for confounding variables using logistic regression.Three latent maternal exposure clusters were identified: a high-risk (6.15 %), a moderate-risk (22.39 %), and a low-risk (71.46 %) cluster. The prevalence of birth defects was 14.08 %, 0.85 %, and 0.52 % for the high-, middle- and low-risk clusters respectively. After adjusting for maternal demographic variables, women in the high-risk cluster were nearly 31 times  more likely to have an infant with birth defects than low-risk women.A high-risk group of mothers in an area with a high risk for birth defects were screened in our study. Targeted interventions should be conducted with women of reproductive age to improve neonatal birth outcomes in areas with a high risk of birth defects. Birth defects are defined as any abnormality in functions, structures, and metabolism of the fetus that is developed in the maternal uterus . Birth dTo date, there have been many epidemiologic approaches to examine the impact of single risk factors, adjusting for confounding variables using generalized linear models , such as folic acid supplementation, maternal delivery age, or alcohol consumption , 6, 7. HUnlike the traditional single factors approaches, latent class cluster analysis (LCCA) is a model-base clustering approach that examines the interrelationships among multiple risk factors and classifies similar objects into groups , 10. PreIn 2009, we conducted a large-scale, retrospective epidemiological survey to assess the prevalence of and likely risk factors for birth defects from 2006 to 2008 in six counties in Shanxi Province, China. Based on these data, the aims of our present study were twofold: 1) to extract different latent maternal exposure clusters based on overall risk factors in the high-risk area using LCCA; and 2) to explore the association between latent clusters and birth defects, screening expectant mothers developing birth defects in high-risk clusters. Identifying individuals in risk clusters characterized by different risk factors will be useful to inform targeted interventions for improving neonatal birth outcomes in areas with a high risk of birth defects.A large-scale, retrospective epidemiological survey of birth defects was conducted in six counties in Shanxi Province, China . The survey was conducted by Population and Family Planning Commission (PFPC) of Shanxi Province, using a three-level stratified random cluster sampling technique –14. ThreMothers were interviewed face-to-face by trained, local investigators going from house to house. Questionnaires designed by PFPC of Shanxi Province were completed by investigators during the interviews. The investigation team in each town comprised 10–12 members , with two town PFPC workers acting as supervisors. Diagnostic information for birth defects was obtained from previous medical records. Unclear birth outcomes were examined by clinical doctors.During the survey, all completed questionnaires were monitored and checked by supervisors and investigators on the same day. When errors and/or missing values were detected, those mothers were re-interviewed. Additionally, 1 % of mothers were reviewed randomly by PFPC of Shanxi Province workers via a telephone interview. Informed consent was obtained from each mother at the beginning of the survey. The Human Research Ethics Committee of Shanxi PFPC reviewed and approved the epidemiological survey and our in-depth following up analysis.The questionnaire was divided into six sections. The first section recorded mothers’ demographic data including age, residence , education, occupation, and annual net income per capita. The second section concerned family history . The third section consisted of a series of Yes/No questions about maternal medical history such as hepatitis, epilepsy, and diabetes. The fourth section covered premarital and pre-conception health guidance. The fifth section investigated maternal exposures during the first 12 weeks of pregnancy, and consisted of six parts, each with a specific series of questions: meat and vegetable consumption, folic acid supplementation, maternal illness in pregnancy, drug use in pregnancy, environmental risk factors in pregnancy and unhealthy maternal lifestyle in pregnancy. Response options for meat and vegetable consumption were: 0, 1–2, or ≥ 3 times per week. Folic acid supplementation had a Yes/No option (“Yes” was selected if supplementation had lasted for at least 3 months within the 3 months before or after the start of pregnancy). Maternal illness in pregnancy and drug use in pregnancy were measured with a series of Yes/No questions. Environmental risk factors in pregnancy and unhealthy maternal lifestyle in pregnancy were measured with a series of questions with response options of never (0), occasionally, or often, with the exception of computer use , and pollution source in area of residence . The final section collected demographic data for offspring. If an infant with birth defects was present, further questions were asked on diagnostic methods and types of birth defects.Statistical analyses were carried out in a three-step process. First, we extracted eight important indicator variables: maternal delivery age, genetic factors, medical history, nutrition and folic acid deficiency, maternal illness in pregnancy, drug use in pregnancy, environmental risk factors in pregnancy, and unhealthy maternal lifestyle in pregnancy. Each indicator variable included multi-risk factor items except for maternal delivery age, and were summed and examined as a “total risk factor score” for data dimensionality reduction: ranging from 0 to N (the number of risk factors in each indicator variable) . The risμk and ∑k in general; therefore an LCCA model for p manifest variables  with K classes  is stated as:Second, LCCA was performed with the eight indicator variables using LatentGOLD 4.5 . LCCA 99, 17, 18yi is an object’s score on a set of manifest variables, πk is the marginal probability of latent class k, and μjk and σjk2 are the mean and variance for manifest variables j in class k, respectively. We estimated means and relative frequencies of classes, which were different from the probabilities of class-specific responses. The posterior probability of assigning respondents to the k class, by Bayes’ theorem, is equal to:where Finally, individuals were classified into different latent classes. Once ≥1-class models were obtained in the exploratory LCCA, goodness-of-fit indicators such as Akaike information criterion (AIC), consistent Akaike information criterion (CAIC), Bayesian information criterion (BIC), and entropy and classification errors were used to determine the model of best fit. In general, smaller AIC, CAIC, BIC and classification errors indicate a better model, whereas the opposite applies for the entropy.Controlling for residence, education, and annual net income per capita, we conducted multivariate logistic regression to explore the association between exposure clusters and birth defects using SPSS Version 17.0 .The average maternal delivery age was 26.3 years. More mothers were rural residents (53.56 %), than urban residents (46.44 %). Most mothers had junior high school education (72.87 %); 1.02 % had no schooling; 14.74 % had elementary school education; 8.66 % had senior high school or technical secondary school education; and, 2.70 % had junior college education or above. Around 11.0 % of mothers had an annual net income per capita of less than 1000 Chinese Yuan (¥); 26.32 % had an annual income of 1000–2000¥; 20.22 % of 4000–8000¥; and, 7.51 % had an annual income of more than 8000¥.We conducted 2- to 5-cluster models. Table n = 2258), but consisted of the highest levels of maternal delivery age, genetic factors, medical history, nutrition and folic acid deficiency, and comparatively high levels of the other four indicator variables. Therefore, Cluster 1 was characterized as the high-risk cluster. Cluster 2  comprised the highest level of illness in pregnancy, drug use in pregnancy, environmental risk factors in pregnancy, unhealthy lifestyle in pregnancy, and the lowest levels of the other four indicator variables, and was considered to be the moderate-risk cluster. Cluster 3 was the largest group , characterized by the lowest levels of indicator variables except for a comparatively high level of nutrition and folic acid deficiency and a moderate level of high maternal delivery age, and was considered the low-risk cluster. Figure Table Further, each participant was assigned to the most likely class based on the known parameters of the three-cluster latent structure. Table We explored the association between birth defects and the identified latent classes. Table Birth defects with serious outcomes and complex maternal exposures are highly prevalent in Shanxi Province. The results of the present study showed that a synthetic effect of overall maternal exposures occurred in areas with a high-risk of birth defects. After extracting three maternal risk clusters using LCCA, mothers were classified into three independent clusters . Each cluster differed in maternal exposures and had a different risk of having an infant with birth defects. The high-risk cluster consisted of the smallest proportion of the study population (6.15 %); however, it had the highest birth prevalence of birth defects at approximately 14.08 %. Furthermore, after adjusting for maternal demographic variables, we found that women in the high-risk cluster were nearly 31 times more likely to have offspring with birth defects compared with the low-risk cluster. This suggests that future government-led, integrated interventions in areas with a high-risk of birth defects may be necessary.Our findings confirmed a significant negative effect associated with all risk factors, and extended the results of single-explanatory risk studies based on a generalized linear model adjusted for confounding variables. The findings in this study also serve as an extension to the results of previous latent class studies based on special dietary factors . Researcχ2 = 7.03, p-value < 0.05). This result is consistent with a research finding conducted earlier in Shanxi Province [The high-risk cluster had the highest prevalence of birth defects, and comprised of mothers with high values in maternal delivery age, genetic factors, medical history, and nutrition and folic acid deficiency. Therefore, those in a high-risk cluster should receive the most attention in terms of implementing interventions targeting behavioral change. Our study found that the rate of birth defects is higher in older maternal age group (≥35 years) (2.02 %) than those in younger maternal age group (<35) (1.40 %)  had a prevalence of birth defects of 0.85 %. This group was characterized by high illness in pregnancy, drug use in pregnancy, maternal hazards in pregnancy and unhealthy lifestyle in pregnancy. Specific intervention measures can be carried out for this group, such as pregnancy health education, reducing illness in pregnancy, careful use of drugs, promoting healthy lifestyles, and avoiding unhealthy exposures in pregnancy.The low-risk cluster (71.46 %) had a prevalence of birth defects of 0.52 %, within an overall lower level of risk factors except for nutrition and folic acid deficiency. Our results were consistent with previous research in Shanxi, which further confirms the importance of nutrition and folic acid use .In our study, annual net income per capita was significantly associated with birth defects. Higher income families were less likely to have birth defects, which is consistent with previous study findings . It is ek latent clusters for each mother [k-means [The population-based retrospective epidemiological survey covered all likely risk factors for birth defects in six counties of Shanxi Province from 2006 to 2008. This was a large sample that reflected the global distribution of birth defects in Shanxi Province. Using a combination of a total risk factor score from an accumulated series of risk factors, and LCCA-extracted synergistic effects of multiple risk factor clusters for dimensionality reduction, our study covered overall maternal exposure factors. Furthermore, as a probability-based method, LCCA has an advantage over distance-based cluster analysis, for example, probability of assignment to [k-means . Our anaHowever, our study had some limitations. First, the population-based data obtained was only a preliminary design for risk factors of birth defects . Some exOur results show synergistic effects based on overall maternal exposures and screening of expectant mothers at high-risk of birth defects in a high-risk area. Mothers in our sample were divided into three distinct groups with different risks of having offspring with birth defects: a high-risk cluster, a moderate-risk cluster, and a low-risk cluster. This will inform the provision of targeted interventions for women of reproductive age in areas with a high risk of birth defects. Additionally, as there is an increase in available electronic medical chart data for women of reproductive age, screening high-risk groups based on overall maternal exposures with a cluster-based approach may be an important direction for future study and targeted interventions, therefore contributing to birth defects control.


It was the aim of our study to evaluate the independent effect of preterm prelabor rupture of membranes (PPROM) as a cause of preterm delivery on mortality during primary hospital stay and significant morbidities in very-low-birth-weight (VLBW) infants < 32 weeks of gestation.Observational, epidemiological study design.Population-based cohort, German Neonatal Network (GNN).6102 VLBW infants were enrolled in GNN from 2009-2012, n=4120 fulfilled criteria for primary analysis  or placental abruption as cause of preterm birth).Multivariable logistic regression analyses included PPROM as potential risk factors for adverse outcomes and well established items such as gestational age in weeks, birth weight, antenatal steroids, center, inborn delivery, multiple birth, gender and being small-for-gestational-age.PPROM as cause of preterm delivery had no independent effect on the risk of early-onset sepsis, clinical sepsis and blood-culture proven sepsis, while gestational age proved to be the most important contributor to sepsis risk. The diagnosis of PPROM was associated with an increased risk for bronchopulmonary dysplasia  but not with other major outcomes.The diagnosis of PPROM per se is not associated with adverse outcome in VLBW infants < 32 weeks apart from a moderately increased risk for BPD. Randomized controlled trials with primary neonatal outcomes are needed to determine which subgroup of VLBW infants benefit from expectant or intentional management of PPROM. Preterm prelabor rupture of membranes (PPROM) occurs in 2–3% of all pregnancies and contributes to 30–40% of preterm births . PPROM iThe initial approach to manage pregnancies complicated by PPROM is based on several parameters assessed upon presentation ,9:gestatTo what extent PPROM and AIS contribute to adverse outcome is not yet fully understood. AIS and PPROM may be innocent bystanders, as “mild” perinatal inflammation may enhance the functional maturation of the preterm immune system or, vice st of January 2009 until 31st of December 2012. The inclusion criteria were as follows: birth weight < 1500g and gestational age ≥22+0 and <32+0 weeks, exclusion criteria: lethal malformations, e.g. trisomy 13 and trisomy 18, and maternal causes of preterm delivery, e.g. placental abruption, HELLP (highly elevated liver enzymes and low platelets) syndrome and pre-eclampsia. After written informed consent was given by the parents, infants were enrolled in GNN by the attending physicians and a predefined clinical data set of 220 parameters including antenatal/postnatal treatment and outcome data were recorded by according data sheets. After discharge, data sheets were sent to the GNN center in Lübeck. A basic data set of other infants born in GNN centers but not enrolled in GNN including birth weight, gestational age, and major outcomes including cause of death was documented. A physician trained in neonatology evaluated the data quality by annual on site monitoring of data sets.We performed an observational study and enrolled VLBW infants in a multi-center trial involving 46 neonatal intensive care units in Germany (GNN). The data were collected from infants born between 1th percentile for gestational age according to gender-specific standards for birth weight by gestational age in Germany [Preterm prelabor rupture of membranes (PPROM) was defined as non-iatrogenic rupture of membranes before the onset of labor effective contractions. Gestational age was calculated from the best obstetric estimate based on early prenatal ultrasound and obstetric examination. Small-for-gestational age (SGA) was defined as a birth weight less than 10 Germany . Amnioti Germany ,18. BlooEarly-onset sepsis (EOS) was defined as blood-culture confirmed sepsis occurring within the first 72 h of life. Late-onset sepsis (LOS) was defined as blood-culture confirmed sepsis occurring ≥72 h of life.Death was defined as all-cause mortality occurring after admission to NICU before discharge home. BPD was diagnosed when needing oxygen at 36 weeks of post menstrual age . IntraceAntenatal steroids was defined as maternal treatment with betamethasone or dexamethasone for lung maturity of the fetus irrespective of number of doses. Antenatal antibiotics and tocolytics were defined as maternal treatment with antibiotics or tocolytics within the time frame of ≤ 24h before delivery.p value < 0.05 was considered as statistically significant for single tests. To determine the independent impact of PPROM on outcomes, we performed multivariable logistic regression analyses including well-established risk factors of adverse short-term outcomes in the whole cohort of infants < 32+0 gestational weeks enrolled in GNN.Data analysis was performed using the SPSS 20.0 data analysis package . Differences between infants born after PPROM and infants without PPROM were evaluated with Chi-square test, Fisher´s exact test and Mann-Whitney U test. A Written informed consent was obtained from parents on behalf of the infants enrolled in our study. The study parts were approved by the local committee on research in human subjects of the University of Lübeck  and the local ethical committees at the other study centers.Specifically: Ethical Board of the Medical Chamber of the North Rhine region, Ethical Board of the University of Aachen, Ethical Board of the University of Bonn, Ethical Board of the Medical Chamber of the federal state of Mecklenburg-Vorpommern, Ethical Board of the Medical Chamber of Berlin, Ethical Board of the University of Magdeburg, Ethical Board of the University of Halle, Ethical Board of the University of Tübingen, Ethical Board of the Medical School Hannover, Ethical Board of the University of Cologne, Ethical Board of the University of Essen, Ethical Board of the Medical Chamber of the Westphalia-Lippe region, Ethical Board of the Medical Chamber of Hamburg, Ethical Board of the Medical Chamber of the federal state of Hessen, Ethical Board of the Medical Chamber of the federal state of Baden-Württemberg, Ethical Board of the Medical Chamber of the federal state of Bavaria, Ethical Board of the Saar University.During the observational study period from January 2009 until December 2012 6102 VLBW infants were enrolled in GNN. 4120 infants fulfilled criteria for primary analysis . As shown in As demonstrated in 2 supplement.No differences were noted with regard to short term outcomes in both groups stratified to PPROM . In TablTo determine whether there is an independent effect of PPROM on the risk of adverse outcomes we tested PPROM along with well-known influencing factors . In our analysis, PPROM had no independent effect on the risk of blood-culture confirmed total sepsis incidence and EOS. In addition to that, PPROM had no influence on total clinical sepsis incidence and clinical sepsis < 72 h . Gestational age proved to be the most important contributor to sepsis risk .As outlined in PPROM had no significant impact on the risk of IVH, PVL and mortality. Major contributors to mortality risk were gestational age, birth weight, gender and multiple birth, while the risk for IVH was associated with multiple birth, gestational age, gender and treatment with antenatal steroids.Our study presents observational data of the large-scale multicenter GNN cohort evaluating the effect of PPROM as cause of preterm delivery on morbidity and mortality in very-low-birth-weight infants < 32 weeks. Our data suggest that PPROM per se is not associated with adverse outcome. We determined a moderately increased risk for development of BPD associated with PPROM which is restricted to infants born 27+0–31+6 weeks.The major strengths of our observation are the large cohort size of a multicenter setting and the accurate phenotypic characterization of infants. This is achieved by regular on-site monitoring of data quality. PPROM management in the most vulnerable gestational age groups is still a matter of debate as conclusive studies are lacking –23. PopuCurrently, the main clinical question whether pregnancies that are aggressively managed differ in infant outcomes than those managed more conservatively still remains unanswered , 22). IdAnother limitation of our study is potential selection bias which is regarded as major challenge in population-based data. For example, NICU mortality in our study cohort is low and does not fully reflect the mortality rate of infants born in GNN centers. Specifically, the study enrolled 69.5% of infants born in centers of GNN . A basic data set including major outcomes (but not cause of preterm delivery) was documented for non-enrolled patients. Notably, the mortality during primary stay in hospital differed remarkably between enrolled patients (4.2%) and non-enrolled infants (25.6%). This is mainly explained by a large proportion of early death in non-enrolled patients due to primary palliative (comfort) care (18.6% of deaths), congenital anomalies (6.2% of deaths) and RDS/early respiratory failure which accounted for 24.6% of NICU deaths. Other less acute causes of death in non-enrolled patients included pulmonary haemorrhage (5.3%), sepsis (15%), NEC/FIP (8.5%), higher grade IVH (10.2%), BPD (2.2%), and others (8.9%).Another important aspect for future design of randomized trials is the correct diagnosis of AIS following PPROM. Prolongation of pregnancy (to avoid morbidity associated with prematurity) in line with prevention of foetal infection is a major challenge in obstetric management of PPROM. Clinicians are thereby faced with a crucial diagnostic problem of relevant AIS/chorioamnionitis/fetal inflammation, Our analysis revealed that 50% of infants born after PPROM were also exposed to AIS. However, only 2.6% developed EOS while 14% had a clinical sepsis within the first 72 hours of life. Infants with PPROM with or without concurrent AIS did not differ remarkably in outcomes in our setting. Furthermore, AIS was not found to be an independent risk factor for major neonatal complications (data not shown) indicating that current diagnostic criteria of AIS are not specific enough to select the most critical cases. With regard to obstetrical management PPROM infants and infants without PPROM had no differences in rates of spontaneous delivery (12.4 vs. 11.0%), elective C/S (79.3 vs. 79.6%) and emergency caesarean section (8.2 vs. 9.3%). But it has to be considered in future studies that remarkable differences between study centers exist, e.g. the spontaneous delivery rate in PPROM patients differed from 0–47%. In addition to that, the study centers themselves are important covariates in determining neonatal outcomes, i.e. risk for EOS, BPD and PVL.Besides center-specific variation there might be selection bias of centers with favourable outcomes in GNN which is a well-known limitation of population-based studies . In addiDespite limitations our large-scale epidemiological data provide a basis for the important interdisciplinary discussion on PPROM-related outcome. The diagnosis of PPROM per se is not associated with adverse outcome in VLBW infants < 32 weeks apart from a moderately increased risk for BPD. Randomized controlled trials with primary neonatal outcomes are needed to determine which subgroup of VLBW infants benefit from expectant or intentional management of PPROM.