N = 27,518), we examined racial/ethnic differences in the modifying effect of CV on the association between paternity status and PTB. Results showed that lack of established paternity was associated with incremental greater odds of PTB across CV quartiles in NH-Whites , NH-Blacks , and Hispanics . Odds of PTB were highest among NH-White women. Public health practitioners should be aware of the negative effect of lack of paternal presence on PTB in women resident in high violence rate communities and racial/ethnic differences that exist.Preterm birth (PTB) is a major public health concern in the US. Lack of established paternity has been linked with increased risk of PTB. Community violence (CV) may modify the association, and racial/ethnic differences may exist. Using a geographically defined cohort of women in Richmond, Virginia ( Preterm birth, defined as the birth of a baby before 37 completed weeks of gestation, is an important cause of perinatal mortality in the US and contributes significantly to perinatal and infant morbidity . About 1For decades, research has focused on identifying maternal risk factors linked to the delivery of a preterm baby. Some of the known risk factors include previous preterm delivery, multiple gestation, underweight and obesity, short and long interpregnancy intervals, tobacco use, bacterial vaginosis, and low socioeconomic status, among others , 9. HoweOne mechanism underlying the association between paternity status and preterm birth is maternal stress. Maternal stress is believed to play a major role in increasing the risk of preterm birth by activating the neuroendocrine system which triggers the release of mediators such as adrenalin, corticotrophin-releasing hormone, cortisol, and other chemical messengers that increase the risk of preterm birth , 16. TheThe Role of Community Violence and Race/Ethnicity. Community violence has been reported to be associated with preterm birth [rm birth , 19–21. rm birth . Given trm birth , 23, it rm birth . ConversDespite the substantial body of literature that have examined the effect of paternity status on poor birth outcomes such as preterm birth , 17, 24,N = 27,518).Live birth records, police crime reports, and census data for Richmond, Virginia, were utilized in this study to create a geographically defined cohort of women. Birth records of all singleton live births were obtained from the Virginia Department of Health Vital Statistics for a 10-year consecutive period (2004–2013). Birth records contained information on maternal sociodemographic history , reproductive history , and risky behaviors . In addition, live birth records included information on presence or absence of the father's first and/or last name. Live birth records were geocoded using maternal addresses using ArcGIS version 10  to identify residential census tracts. About 89 percent of the birth records were successfully geocoded. The live birth records were then linked to the 2010 US census data using the unique residential census tract identifiers described above. Crime reports were obtained from the Richmond Police Department, Virginia. The police crime report provided data on crime incidents which occurred during the 10-year period from 2004 to 2013 in Richmond, Virginia. Crime data included all Class A reportable offenses  involving youths aged 10–24 years. The mean incidence rate of violence was calculated for each census tract for the 10-year period and merged with the geocoded live birth record. The final data had a 2-level hierarchical structure with live births nested in 64 census tracts. Detailed description of the data has been reported elsewhere . The stuOutcome. Preterm birth was defined as the birth of a baby before 37 completed weeks of gestation. Gestational age at birth was categorized into 2 levels: preterm birth (<37 weeks) and term births (≥37 weeks). Gestational age was determined using the clinical estimate computed by the physician. In situations where the clinical estimate of gestational age was missing or implausible (<1%), gestational age was computed by taking the interval between date of last menstrual period reported by the mother at her first prenatal visit and the date of delivery.Exposure. Paternity status was measured as a binary variable (paternity established or paternity not established) using the presence or absence of the father's first and/or last name on the birth certificate, as has been used previously in the literature [terature , 24. Theterature .Effect Modifiers. Community violence was measured at the census tract level. It was obtained by aggregating Class A reportable offenses  involving youths aged between 10 and 24 years to produce a single violence variable. Violence rates were calculated for each census tract and the mean violence rate per 1,000 youth population during the 10-year period was computed. The mean violence rate was then categorized into quartiles, ranging from low (quartile 1) to high (quartile 4).Maternal race/ethnicity was categorized as NH Whites, NH Blacks, and Hispanics. Other NH racial/ethnic groups such as Asians, American Indians, Hawaiians, and Alaskan Natives were excluded due to small numbers.Covariates. Factors examined include individual- and community-level factors. Individual-level factors include maternal age , maternal education , insurance , tobacco use (yes or no), alcohol drinking (yes or no), previous preterm birth (yes or no), parity , and adequacy of prenatal care . Prenatal care adequacy was defined using the Kotelchuck index which takes into account when prenatal care was initiated and how many visits were completed, controlling for estimated gestational age at delivery [delivery . Communi∗ community violence ∗ race/ethnicity) to the unadjusted model and statistically tested using the likelihood ratio test (p = 0.0134). As recommended in the literature [∗ community violence, paternity status ∗ race/ethnicity, and community violence ∗ race/ethnicity) were also included in the model. The best fitting model was thus stratified across quartiles of community violence and race/ethnicity. The conceptual model is shown in All analyses were conducted using SAS 9.4 . Descriptive analyses were conducted to determine the characteristics of the study population across individual- and community-levels. The GLIMMIX procedure in SAS was then used to fit four multilevel logistic regression models for the response variable (preterm birth) using a binary distribution and logit link function. Model 0 was an unconditional model . This will facilitate the calculation of the intraclass correlation coefficient (ICC) to estimate how much variability in the response variable is explained by the individual and community-level factors. Model 1 (unadjusted model) examined the association between paternity status and preterm birth. Model 2 was fit to control for individual-level factors. Model 3 controlled for individual-level factors from model 2 in addition to community-level factors. Model fit was assessed using the deviance test utilizing Laplace estimation and multiple testing correction was done using the Bonferroni correction (0.05/4 = 0.0125). To explore racial/ethnic differences in the modifying effect of community violence on the association between paternity status and preterm birth, community violence and race/ethnicity were entered as a three-way interaction term  reported no established paternity. A greater proportion of NH Black women, adolescents and young adults, women on Medicaid, high school graduates or less, women with parity of 2 or higher, and women with inadequate prenatal care were without established paternity (p < 0.0001). For community-level factors, a greater proportion of women in the upper quartiles of community violence reported no established paternity (<.0001) and on average, a greater proportion of female-headed households, households below 100% of the Federal Poverty Level, and NH Black population reported no established paternity (<.0001).p = 0.0297) amount of variability in the log odds of preterm birth between the census tracts, and so community-level factors were retained. This was consistent with the deviance tests.Multilevel regression analyses of paternity status and preterm birth stratified by community violence and race/ethnicity for the best fitted model (model 3) are shown in This study shows that lack of established paternity is associated with greater odds of preterm birth in women, and the odds of preterm birth increase with higher rates of community violence for NH White, NH Black, and Hispanic women. These findings support our study hypothesis and are consistent with previous studies that reported greater risk of preterm birth in women with no established paternity , 28. TheFurthermore, the odds of preterm birth in women who had births with no established paternity were shown to increase with increasing rates of community violence. This finding was not surprising and is consistent with our study hypothesis, as previous studies have reported that the risk of preterm birth increases with higher rates of community violence , 21. It Another important finding observed in this study was the attenuation of the odds of preterm birth in women who had births with no established paternity and were exposed to the highest rates of community violence (quartile 4), in both NH White and NH Black women. This finding was in contrast to previous studies that have shown that women in areas with high rates of community violence have the greatest risk of preterm birth , 21. TheThis study utilized a geographically defined cohort of women with singleton live births in Richmond, Virginia, USA, which has one of the highest crime rates in the US ; hence, This study demonstrated that lack of established paternity is associated with increased odds of preterm birth, and the odds of preterm birth increased as the rate of community violence increased in NH White, NH Black, and Hispanic women. Community violence was also shown to have differentially greater modifying effect on the odds of preterm birth in NH White women who had no established paternity, relative to NH Black and Hispanic women. Healthcare providers and policy makers should be aware of the negative effect of lack of paternal presence on preterm birth in women resident in communities with high rates of violence and the racial/ethnic differences that exist. Intervention programs to provide support to pregnant women with no paternal presence who are resident in communities with high rates of violence are needed. Receipt of social support may help women to better cope with stressors that may arise from lack of paternal presence and exposure to community violence .


Elegance is a prized quality in science that is associated with simplicity and explanatory power. This essay considers the qualities that make a scientific model, experiment, method, or theory “elegant,” with a focus on the life sciences. We propose a definition of elegance that includes clarity, cleverness, correctness, explanatory power, parsimony, and beauty. The pursuit of elegance can improve the quality of science, but elegance must be pursued with caution, as the truth is sometimes inelegant. Elegance should be left to shoemakers and tailors.—Ludwig BoltzmannNew Yorker  that was derived from the Latin elegantem (Merriam-Webster Dictionary defines “elegant” as “tasteful richness of design or ornamentation, dignified gracefulness or restrained beauty of style,” and in the specific context of science, as exhibiting “precision, neatness, and simplicity” (The word “elegant” entered the English language in the 1400s, a French word (legantem . The Merplicity” .60 molecule buckminsterfullerene noted that an elegant theory or model must explain a phenomenon “clearly, directly and economically” (New Yorker article adds that elegance in science requires “simplicity plus capaciousness” in explanatory power (Glynn has argued that the essence of scientific elegance is simplicity and explanatory power, while at the same time noting that its appreciation requires a historical context . A recenmically” . The Newry power . This foThe question of whether elegance is merely desirable or essential for good science is particularly relevant to current concerns about rigor and reproducibility in biomedical research . If elegA search of the PubMed database for the word “elegant” yields the titles and abstracts of more than 4,000 publications. Browsing through those publications reveals that the word “elegant” is used to describe models, experiments, methods, or theories. Furthermore, the appearance of the word “elegant” in PubMed publications is increasing almost linearly over time. In the mid-1990s, fewer than 100 articles used the word “elegant” in titles and abstracts, whereas in recent years the number of articles using this word now routinely exceeds 300 per year . Hence, The structure of DNA is often referred to as an elegant model. In 1953, Watson and Crick reported that the structure of DNA is a double helix . PaulingThe Watson-Crick model made the testable prediction that DNA replication would be semiconservative, providing the basis for the Meselson-Stahl experiment. In 1958, Meselson and Stahl isotopically labeled DNA bases and separated the products by ultracentrifugation to show that each new strand of DNA is built upon a previously existing strand . This waElegance also applies to the method of PCR developed by Kary Mullis to amplify DNA from minute quantities of single-stranded template , 21. PCRThe Origin of Species (The most elegant theory in the biological sciences is unquestionably Darwin’s theory of evolution by natural selection, which was all the more remarkable for being proposed before genes or DNA were known. As with PCR, some of the basic principles underlying natural selection were proposed well before Darwin published  Species ; new speThe theme of simplicity runs deep in elegant science, and definitions of elegance invariably contain such words as “simplicity,” “neatness,” and “economy.” However, it is not immediately obvious that simplicity should be considered necessary or sufficient for scientific elegance. Elegance does not connote simplicity in fashion, as it is hard to imagine a bikini being viewed as elegant. The association of simplicity with elegance in science may relate to the principle of parsimony also known as Occam’s razor, which states that the simplest explanation is the most likely to be correct. This in turn implies that for science to be elegant, it must also be correct. What Occam, a 14th century theologian, actually wrote was “Non sunt multiplicanda entia sine necessitate,” which can be translated as “Entities must not be multiplied beyond necessity.” This principle is not without risk. As Francis Crick observed, “Occam’s razor … can be a very dangerous implement in biology. It is … very rash to use simplicity and elegance as a guide in biological research” . He alsobut no simpler.” Hence, there are limitations on the relationship between simplicity and elegance in science.Nevertheless, we suggest that parsimony is preferable to simplicity, as simplicity in science is relative. For example, when some physicists expressed frustration over the complexity of Einstein’s theories of relativity in comparison to Newton’s formulas, he responded that “It can scarcely be denied that the supreme goal of all theory is to make the irreducible basic elements as simple and as few as possible,” but he cautioned that this must be achieved “without having to surrender the adequate representation of a single datum of experience” . This inFrom the foregoing discussion, we propose that to be considered elegant, science should meet the five criteria of being clear, clever, correct, explanatory, and parsimonious . This coAlthough scientific rigor  is neces


Geostationary Operational Environmental Satellite (GOES) classes, with an emphasis on including powerful confined flares and weak eruptive flares. HMichelson Doppler Imager (MDI) and Helioseismic and Magnetic Imager (HMI) magnetograms were used to derive the ribbon separation, the ribbon-separation velocity, the magnetic-field strength, and the reconnection electric field. We find that eruptive flares reveal statistically larger ribbon separation and higher ribbon-separation velocities than confined flares. In addition, the ribbon separation of eruptive flares correlates with the GOES SXR flux, whereas no clear dependence was found for confined flares. The maximum ribbon-separation velocity is not correlated with the GOES flux, but eruptive flares reveal on average a higher ribbon-separation velocity (by ≈ 10 km s−1). The local reconnection electric field of confined  contains supplementary material, which is available to authorized users. The set covers 50 events in total, ranging from GOES classes B to > X10, including 19 eruptive and 31 confined flares.The data set consists of 50 Hsidc.be/educational/classification.php#OClass) 4 and 3. Then an appropriate number of importance class 2 flares were selected. For the importance class 1 and S flares, we searched for suitable flares beginning from 2015 and going backward in time. In addition, we placed an emphasis on including powerful confined as well as weak eruptive flares. The flares were selected to be close to the center of the solar disk (CMD < 45°), in order to minimize projection effects. The central meridian distance (CMD) is the angular distance in solar longitude measured from the central meridian. We aimed at having good coverage over Hmentclass2pt{minimSolar and Heliospheric Observatory/Large Angle and Spectrometric Coronagraph (SOHO/LASCO) CME catalog (cdaw.gsfc.nasa.gov/CME_list) was checked to determine the flare–CME association. The flare position had to be consistent with the position angle given in the CME catalog and the flare had to occur within 60 minutes of the linearly extrapolated starting time of the CME. For the M1.2/1N flare on 1 October 2011, an eruption in the original LASCO movie can be seen, but no entry in the SOHO/LASCO CME catalog exists. We therefore refer to Temmer et al.  up to ≈ 70 V cm−1 for the most powerful flare under study (X17.2/4B flare on 28 October 2003), covering almost two orders of magnitude.Figure Figures entclass1pt{minimavis-à-vis each other are represented by filled symbols. Figure i.e. the strength of the flare , whereas the majority of confined flares tends to have maximum ribbon-separation speeds < 30 km s−1.The maximum ribbon-separation velocity of eruptive flares shows a wide range  and eruptive (38%) events, distributed over GOES classes B to > X10. We analyzed flare events that occurred from June 2000 to June 2015, homogeneously covering all HSu, Golub, and Van Ballegooijen  studied cf. Figure e.g., Thalmann et al., However, the ribbons of about 70% of confined flares do not separate farther than 10 Mm, which is in good agreement with the findings of Kurokawa  and Su, et al.  were obtained at flare-ribbon locations where HXR footpoints are located, and the weakest electric fields (≈ 3 V cm−1) were found in regions without HXR sources. For the X3.8/3B flare on 17 January 2005, we obtain ≈ 40 V cm−1. Comparing the two tracking directions in Temmer et al.  surrounding are known to also control the eruptive behavior of flares . Comparing the fluxes for two height regimes (1.0 – 1.1 R⊙ and 1.1 – 1.5 R⊙), they found that the ratio of the horizontal flux in the low corona divided by that in the high corona was significantly higher for eruptive flares. The theoretical work by, e.g., Török and Kliem  and the Ramaty High Energy Solar Spectroscopic Imager (RHESSI) have been performed where magnetic reconnection could directly be observed  including magnetographs at all spacecraft may provide a great step forward in better determining the governing physical processes from the observations.In order to shed more light on the reconnection process of solar flares, combined measurements from spacecraft at different positions in the heliosphere would be helpful. Recently, case studies using the (MP4 434 kB)(MP4 271 kB)(MP4 2.8 MB)Below are the links to the electronic supplementary material.


The human brain anatomy is characterized by dramatic structural changes during fetal development. It is extraordinarily complex and yet its origin is a simple tubular structure. Revealing detailed anatomy at different stages of brain development not only aids in understanding this highly ordered process, but also provides clues to detect abnormalities caused by genetic or environmental factors. However, anatomical studies of human brain development during the fetal period are surprisingly scarce and histology-based atlases have become available only recently. Diffusion tensor imaging (DTI) measures water diffusion to delineate the underlying neural structures. The high contrasts derived from DTI can be used to establish the brain atlas. With DTI tractography, coherent neural structures, such as white matter tracts, can be three-dimensionally reconstructed. The primary eigenvector of the diffusion tensor can be further explored to characterize microstructures in the cerebral wall of the developmental brains. In this mini-review, the application of DTI in order to reveal the structures of developmental fetal brains has been reviewed in the above-mentioned aspects. The fetal brain DTI provides a unique insight for delineating the neural structures in both macroscopic and microscopic levels. The resultant DTI database will provide structural guidance for the developmental study of human fetal brains in basic neuroscience, and reference standards for diagnostic radiology of premature newborns.


Intrauterine growth restriction affects 3% of newborns; and the lightest 10% of whom are classified as small for gestational age (SGA). These low-birth weight newborns are at increased risk of neonatal morbidity such as hypoxia and hypoglycaemia. In later life, they are at higher risk of several age-related diseases such as cardiovascular and metabolic disorders and dementia. As having short telomeres is also associated with these diseases, we tested if these newborns might already start with shorter telomeres at birth.Relative telomere lengths were determined using quantitative real-time PCR in cord blood samples from 195 newborns of Chinese ancestry. Based on the telomere length normalised to a single copy gene and a reference DNA sample as internal control, we found statistically significant correlations between relative telomere length and both unadjusted and gestational age-adjusted birth weight, with the lighter newborns having shorter telomeres. The SGA birth weight group comprising the bottom 10% of the samples also had the shortest telomeres compared to the medium and heaviest birth weight groups.Our results indicate that there is reduction of cord blood telomere length for newborns with lower birth weight. Telomeres are highly conserved hexamer repeats (TTAGGG) at the ends of all chromosomes . They maThe enzyme telomerase which maintains and restores the terminal regions is only expressed in embryonic/foetal cells, germ cells and other proliferating organs in adults. Short telomeres are associated with genome instability, cellular aging, some types of cancer, and several age-related diseases such as heart disease, atherosclerosis and diabetes , 5. AlthSeveral methods have been developed to assess telomere length. They include procedures which quantify the mean from total DNA extracted from a population of cells, and those which measure the telomere of individual chromosomes in a cell . The mosWith the advent of real-time polymerase chain reaction platforms, a method based on quantitative real-time PCR (qPCR) which requires about 100× less DNA, has higher throughput and faster turnaround time has become more commonly used . The meaIn this study, we tested the hypothesis that cord blood telomere length might be directly related to birth weight. Our objective was to determine the relationship between birth weight and cord blood telomere length in newborns of Chinese descent that were born in our hospital. We measured the telomere length of these cord blood samples using qPCR to assess correlation with birth weight, taking into account gestational age and maternal age.The study protocol was approved by the SingHealth Institutional Review Board which oversees all research activities in the hospital. Inclusion criteria were deliveries from mothers of Chinese descent from singleton pregnancies with no recognizable congenital disorders. Approval was granted for the anonymous collection of cord blood and limited demographic data without requiring written informed consent.Cord blood samples were collected between March 2011 and April 2012 in our hospital, for cases that met inclusion criteria and were not involved in other research studies requiring cord blood, had not consented to donate to cord blood bank, or had not planned to store it with a private operator. Data on gestational age, gender and birth weight of the newborns, and maternal age were anonymously collected.Genomic DNA was extracted from the cord blood of 195 Chinese newborns  using Puregene Blood Kit . The quantity and quality of extracted DNA was determined by using a NanoDrop 1000 spectrophotometer .36B4 as the reference gene for normalisation using published primer sequences [36B4 was made up of 15 ng genomic DNA, 1× Roche FastStart Universal SYBR Green Master (ROX) , 300 nM of the forward and 500 nM of the reverse primers while the telomere reaction mix had 270 nM of the forward and 900 nM of the reverse primers. Water was added to a final volume of 20 μl. DNA samples from two healthy females were included in every experiment as internal calibration to control for inter-experimental variability. Amplification for each sample was done in quadruplicates in the Applied Biosystems StepOnePlus real time PCR system  using the following cycling condition: 50 °C for 2 min and 95 °C for 10 min, followed by 40 cycles of denaturation at 95 °C for 15 s, and annealing-extension at 56 °C for 1 min and dissociation curve analysis.Telomere length assessment was done by qPCR, with equences . AmplifiT values were obtained by setting the threshold at the exponential phase of amplification. The average CT values of the best three of the quadruplicates were used with one set of data discarded. Telomere copy number relative to single copy gene (T/S) ratio was calculated from the CT values using the following formula: Relative T/S = 2–(∆CT Sample − ∆CT control sample) = 2−(∆∆CT). Where ∆CTSample = Sample CT(tel) – Sample CT(36B4) and ∆CTcontrol = Control CT(tel) – Control CT(36B4). Relative telomere length (RTL) was then obtained by comparing the T/S ratio of the test sample relative to an internal control DNA sample included in every experiment to correct for any batch variation for assays done at different times. The coefficient of variation for RTL values obtained using Control #1 is 12.7% and for Control #2 is 12.2%.Result was analysed using the Applied Biosystems StepOne™ software (version 2.1). CP value of <0.05 was considered as statistically significant. Quantitative variables are expressed as mean and standard deviation. For each sample, RTL used was the average of the best three of the four replicates relative to each of the two control DNA samples, and gestational age-adjusted birth weight was generated by dividing the birth weight by gestational age. Correlation between individual quantitative variables was evaluated using a Pearson test. The relationship between RTL and birth weight was also explored while controlling for maternal age using partial correlation.All analyses were carried out using IBM Statistical Package for the Social Sciences version 19 . A http://www.msdmanuals.com/professional/pediatrics/perinatal-problems). Age-adjusted birth weight was used to classify the samples: lightest 10% (n = 20) in the small for gestational age group (SGA), heaviest 10% (n = 20) in the large for gestational age group (LGA), and the rest (n = 155) in the appropriate for gestational age group (AGA). Differences between the three groups were assessed using ANOVA for quantitative variables and Chi Square for categorical variables.The newborns were placed into three groups based on clinical definition that is generally used to categorise extreme fetal growth groups  and gestational age from 32 to 41 weeks (mean 38.15 ± 1.23 weeks). Of the 195 samples with DNA available for telomere length qPCR analysis, there were 97 males and 98 females. The experimentally determined telomere length showed normal distribution  normalised to Control #1 , with the mean for Group AGA at 0.8270 (P = 0.050). The difference between Group AGA and Group LGA was not statistically significant.Analysis using tertile birth weight groups and different models  did not reveal any statistically significant correlation. The extreme group design was then used to classify the newborns into three groups based on age-adjusted birth weight: lightest 10% (Group SGA), heaviest 10% (Group LGA), and the rest (Group AGA). There were statistically significant differences among the groups in terms of RTL P = 0.02 normaliP = 0.009) in mean gestational age for the three groups: with the mean for SGA at 37.30 ± 2.27 weeks, AGA at 38.24 ± 1.03 and LGA at 38.30 ± 0.865. The median for all three groups were the same at 38.00 weeks. There was no statistically significant difference between the three groups in the mean for maternal age (P = 0.653) or TRF measured previously (P = 0.332).There was a statistically significant difference (P = 0.153). There was also no statistically significant difference in the mean maternal age between the preterm and full-term birth groups (P = 0.911).Eight cases were from preterm deliveries of between 32 and 36 weeks . Although the preterm group has shorter RTL of 0.933 ± 0.139 compared to the 0.998 ± 0.125 for the full-term group , the difference was not statistically significant  placentas compared to controls as measured by quantitative FISH , 18. OurAdditional analysis using tertile birth weight groups and different models did not reveal any statistically significant correlation. Conversely, there is a statistically significant difference when the newborns are divided into three groups based on gestational age-adjusted birth weight, with the shortest telomeres found in the group with the lowest birth weight. This extreme group method of categorizing birth weight (SGA/AGA/LGA) is typically used by studies investigating the association of telomere length with birth parameters. It is based on clinical relevance as both LGA and SGA groups are at higher risk of maternal health complications, adverse neonatal outcomes and future health morbidity. LGA newborns are at increased risk of birth injuries during deliveries. They may have polycythemia, hypoglycemia and immature lungs if the mothers are diabetic, and increased risk of diabetes and obesity in adulthood . HoweverFor SGA neonates and in particular, those with intrauterine growth restriction (IUGR), there is an increased risk of cardiovascular and metabolic diseases . Some ofA recent study (N = 69) found an inverse correlation between telomere length and birth weight groups, with the LGA group having the shortest telomeres and the SGA group having the longest compared to their average weight counterparts . In our We previously reported that females have longer telomeres as measured by TRF. The data for qPCR RTL showed a similar trend although the difference was not statistically significant Table . RecentlThere are several limitations of this study. We do not have the clinical data on the mothers who might have medical conditions that impacted the growth of their foetuses; our data on birth weight and gestational age was collected anonymously and the mothers’ pre-pregnancy weight or body mass index was not available for analysis. We also did not have the DNA from the parents for parental RTL measurement for correlation or adjustment of the newborns’ RTL data. In addition, the qPCR method measures the average TL of all chromosomes in multiple cells, which cannot distinguish whether some cells or chromosomes have some telomeres that are significantly shorter or longer than the rest.In summary, our RTL results using a medium sized cohort from a single ethnic group show a direct correlation of birth weight with mean qPCR telomere length which measures the true length of the telomere hexamers. As telomere length is a reflection of replicative history, the shorter telomeres could be a reflection of higher cellular turnover during the embryonic period for the newborns that had suboptimal growth. Future work could look into whether there might be a subpopulation of cells with shorter telomeres. The presence of individual chromosomes with much shorter telomeres can be explored using more sensitive methods that directly measure telomere length at the resolution of a single chromosome end.


Birth registration, and the possession of a birth certificate as proof of registration, has long been recognized as a fundamental human right. Data from a functioning civil registration and vital statistics (CRVS) system allows governments to benefit from accurate and universal data on birth and death rates. However, access to birth certificates remains challenging and unequal in many low and middle-income countries. This paper examines wealth, urban/rural and gender inequalities in birth certificate coverage.We analyzed nationally representative household surveys from 94 countries between 2000 and 2014 using Demographic Health Surveys and Multiple Indicator Cluster Surveys. Birth certificate coverage among children under five was examined at the national and regional level. Absolute measures of inequality were used to measure inequalities in birth certificate coverage by wealth quintile, urban/rural residence and sex of the child.Over four million children were included in the analysis. Birth certificate coverage was over 90% in 29 countries and below 50% in 36 countries, indicating that more than half the children under five surveyed in these countries did not have a birth certificate. Eastern & Southern Africa had the lowest average birth certificate coverage (26.9%) with important variability among countries. Significant wealth inequalities in birth certificate coverage were observed in 74 countries and in most UNICEF regions, and urban/rural inequalities were present in 60 countries. Differences in birth certificate coverage between girls and boys tended to be small.We show that wealth and urban/rural inequalities in birth certificate coverage persist in most low and middle income countries, including countries where national birth certificate coverage is between 60 and 80%. Weak CRVS systems, particularly in South Asia and Africa lead rural and poor children to be systematically excluded from the benefits tied to a birth certificate, and prevent these children from being counted in national health data. Greater funding and attention is needed to strengthen CRVS systems and equity analyses should inform such efforts, especially as data needs for the Sustainable Development Goals expand. Monitoring disaggregated data on birth certificate coverage is essential to reducing inequalities in who is counted and registered. Strengthening CRVS systems can enable a child’s right to identity, improve health data and promote equity.The online version of this article (doi:10.1186/s12939-017-0635-6) contains supplementary material, which is available to authorized users. A birth certificate is proof of identity, age, and family relationships, and confirms that a child’s birth has been registered. Provided to a child by a civil registry, it allows an individual to make claims of nationality, benefit from government schemes, open a bank account, travel, and vote. Although a birth certificate does not guarantee protection, it can help protect children from abuse and exploitation , 2, reduBirth registration is included in the Sustainable Development Goals (SDGs), which have an explicit aim to ensure access to widely accepted, robust identity credentials, and improve vital registration systems , 10. In In 2013, a UNICEF report showed that socioeconomic status, religion, maternal education, and access to a health facility can determine which children benefit from identification documents . HoweverWe analyzed birth certificate and birth registration coverage through nationally representative household surveys, including 48 Demographic Health Surveys (DHS)  and 46 MWe included all surveys with data on birth registration after the year 2005 . We selected the most recent survey for each country. All surveys were publicly available and ethical clearance was the responsibility of the institutions that administered the surveys. Countries were categorized by region based on UNICEF regions , and by Outcomes were constructed using one standard question from DHS and three standard questions from MICS. All outcomes were based on a caregiver report of whether a child had been registered and whether a birth certificate was issued. Although there were small differences between the DHS and MICS questions, both surveys allowed each child to be classified into three groups: “birth registered, with a certificate (which the caregiver is in possession of)”, “birth registered, no certificate”, or “birth not registered”. Details of the questionnaires are available in Additional file We calculated birth certificate coverage as the primary outcome, defined as the percentage of children under age five (0 to 59 months) with a birth certificate at the time of the survey. The denominator was the population of children under-five surveyed. Since survey sampling was nationally representative, our estimates represent national coverage in each country. Two secondary outcomes were also used in select analyses: first, the proportion of children under five who were reported to have been registered but did not receive a certificate; and second, birth registration coverage, which was calculated as the percentage of children under age five whose birth was reported as registered – this included children with and without a birth certificate. Children under five who were not included in these categories were classified as unregistered.National estimates were compared to published MICS and DHS reports. All differences between calculated and published estimates were smaller than one percentage point, with the exception - as expected - of the nine MICS 3 surveys included in the analysis due to the standardization of skip patterns between MICS 3 and MICS 4/5.We disaggregated the outcomes in each survey by sex of the child, wealth quintile and urban or rural residence. The wealth index is calculated through principal component analyses of household assets and building characteristics in each country; the first quintile represents the poorest 20% of all households, and the fifth quintile represents the wealthiest 20% of households in the sample in each country –18. The We calculated point estimates, standard errors and confidence intervals for each outcome, in each country.n = 3) and countries which were not assigned to an income group (n = 2), were excluded from income group estimates.Unweighted averages of country estimates were calculated for each UNICEF region and for each World Bank income group at the time of the survey. High income countries . Using logistic regression, the SII represents the absolute difference in the fitted value of birth certificate coverage between the highest and the lowest extremes of the wealth distribution , 19. TheAll country-level analyses accounted for the multi-stage survey design, including sampling weights and clustering. Analyses were carried out in Stata†/MP13 .Data on birth certificate coverage in 94 countries from seven UNICEF regions were included in the analysis. This included a total of 4,195,238 children. Nineteen surveys were conducted before 2010, and 75 surveys were conducted between 2010 and 2014. Data on World Bank income group at the time of the survey were available for 92 countries  of which 32 were low-income countries, 40 were lower-middle income countries, and 17 were upper-middle income countries. The 3 high-income countries were removed from the income group analyses. Disaggregated coverage estimates by sex were available for all 94 countries, and by urban/rural residence and wealth for 92 countries.Birth certificate and registration coverage varied by UNICEF region and World Bank income group Fig. . EasternBirth certificate coverage was over 90% in 29 countries - of which coverage was over 98% in 16 countries - and below 50% in 36 countries  in Eastern & Southern Africa. However, West & Central Africa had the largest gap in coverage, which was 35.7 percentage points lower among children in the poorest quintile compared to children in wealthiest quintile. A similar disadvantage for poor children was prevalent in Eastern & Southern Africa (24.4 percentage points), South Asia (21.7 percentage points), and East Asia & Pacific (22.3 percentage points). Although gaps in coverage by wealth were smaller in Latin America and the Caribbean (11 percentage points) and Central and Eastern Europe (7.2 percentage points), wealth inequalities persisted.Additional file Figure Our analyses revealed large and consistent disadvantages to rural compared to urban children. Regional estimates showed the largest average gap was in West and Central Africa: 22.1 percentage points. Differences were smaller in other regions: 2 percentage points in Latin America and the Caribbean and 4 percentage points in Central and Eastern Europe. The gap was largest in low-income (16.5 percentage points) and smallest in upper-middle income countries (2 percentage points).Country level estimates Additional file 2:Equity analyses and reference groups used in the analysis. (DOCX 53 kb)Additional file 3:Birth Registration Status of Children Under Age Five in 94 countries: country level estimates, sample size, and data sources. (XLSX 27 kb)Additional file 4:Slope Index of Inequality in Birth Certificate Coverage by Wealth Quintile Among Children Under Five. (DOCX 991 kb)Additional file 5:Absolute Difference in Birth Certificate Coverage between Urban and Rural Residence Among Children Under Five. (DOCX 755 kb)Additional file 6:Absolute Difference in Birth Certificate Coverage between Boys and Girls Under Five. (DOCX 660 kb)


Bacteroides in vaginally delivered infants. We aimed to determine the impact of birth mode of the preterm gut microbiome over the first 100 days of life and following neonatal intensive care unit (NICU) discharge. In total, 867 stool samples from 46 preterm infants , median gestational age 27 weeks, were sequenced . Of these, 776 samples passed quality filtering and were included in the analysis. The overall longitudinal alpha-diversity and within infant beta-diversity was comparable between cesarean and vaginally delivered infants. Vaginally delivered infants kept significantly more OTUs from 2 months of life and following NICU discharge, but OTUs lost, gained, and regained were not different based on birth mode. Furthermore, the temporal progression of dominant genera was comparable between birth modes and no significant difference was found for any genera following adjustment for covariates. Lastly, preterm gut community types (PGCTs) showed some moderate differences in very early life, but progressed toward a comparable pattern by week 5. No PGCT was significantly associated with cesarean or vaginal birth. Unlike term infants, birth mode was not significantly associated with changes in microbial diversity, composition, specific taxa, or overall microbial development in preterm infants. This may result from the dominating effects of NICU exposures including the universal use of antibiotics immediately following birth and/or the lack of Bacteroides colonizing preterm infants.The short and long-term impact of birth mode on the developing gut microbiome in neonates has potential implications for the health of infants. In term infants, the microbiome immediately following birth across multiple body sites corresponds to birth mode, with increased  Lactobacillus) and cesarean delivered neonates harboring skin-like microbes   . In the ococcus) . In infa of life . Contrar of life . DiffereIn preterm infants, studies directly exploring associations between birth mode and the temporal microbiome are lacking, although evidence suggests other factors such as feeding practices, postnatal age, and diseases like sepsis or necrotizing enterocolitis (NEC) likely have a bigger association with the microbiome . A recenIn the current study we combined our existing publically available datasets  and our Ethical approval was obtained from the County Durham and Tees Valley Research Ethics Committee. Parental written informed consent was given.® (Bifidobacterium bifidum-ATCC15696 and Lactobacillus acidophilus-NCIMB701748) soon after initial introduction of feeds, where half an Infloran capsule was given twice daily (125 mg b.d. at 10 9 organisms per dose). Infants contributing a minimum of seven samples in their first 100 days were included in the study.The study design, setting, participants, and methods of data collection have been reported previously . BrieflyTM PowerSoil® DNA Isolation Kit  in accordance with the manufacturer’s instructions. The V4 region of the 16S rRNA gene was amplified by PCR using barcoded Illumina adapter-containing primers 515F and 806R  algorithm  to overcome bias of repeated measures in longitudinal analyses. At a given time point, samples within ±10 week days included, where the sample nearest to the specific week were chosen, giving preference to samples collected prior to the time point. Significance of categorical variables were determined using the non-parametric Mann–Whitney test for two category comparisons or the Kruskal–Wallis test when comparing three or more categories . All P-vlgorithm . Linear Table 1). All infants received at least 48 h of antibiotics immediately following birth and the total number of days of antibiotic treatment while on the NICU was comparable between birth modes. Due to a change in unit practice, probiotics were administered in 4/21 of the cesarean infants born after 2013 and no vaginal infants received probiotics (P = 0.07). Use of probiotics was associated with increased relative abundance in only the genera contained within the probiotic , but not in other taxa  mapped to the database, with 776 (760 NICU and 16 post discharge) samples remaining in the analysis following rarefication to 4300 reads. Of these 46 infants, 25 infants were delivered vaginally and 21 infants were delivery by cesarean section  . The Shannon diversity increased from around 0.75 in the days following birth to 1.25 at NICU discharge, with comparable development between cesarean and vaginally delivered infants  . The observed OTUs and Shannon diversity continued to increase following discharge but no significant difference between cesarean and vaginally delivered infants occurred in post discharge samples   and post discharge   than those delivered by cesarean . The number of OTUs kept in month 1 of life was comparable between birth modes (P = 0.947). The microbiome stabilized rapidly from birth to week 4 of life, where the number of ‘OTUs lost’ and ‘new OTUs gained’ declined, but the ‘OTUs regained’ (previously present but not in the preceding sample) increased . However, there was no difference in birth mode between OTUs lost, regained, or newly gained during the first 100 days while on the NICU. Following discharge, vaginally delivered infants continued to have significantly (P = 0.021) increased kept OTUs compared to cesarean, relative to the last NICU sample collected , Escherichia , Enterococcus (15%), Staphylococcus (14%), and Bifidobacterium (5%) dominated NICU samples, accounting for 84% of the total relative abundance in the first 100 days of life . Klebsiella and Enterococcus remained relatively consistent during the NICU period, Staphylococcus and Escherichia declined from birth, and Bifidobacterium gradually increased through the NICU period. To determine significant differences in the relative abundance of genera through the NICU, while accounting for repeated measures, the first sample from each infant in weeks 1, 3, 5, 8, and 10+ were included. This cross-sectional comparison showed no significant difference between cesarean and vaginal infants in any genera at any time point . With the exception of PGCTs 1 and 4, all PGCTs had significantly different Shannon diversity, with PGCTs 2 and 4 showing higher Shannon diversity and PGCT 6 (Staphylococcus dominant) showing the lowest Shannon diversity . Birth mode showed moderate differences in the PGCTs detected over the initial weeks of life . Most notable was the increase of PGCT 3 in cesarean delivery and PGCT 4 in vaginal delivery at week 1 of life. PGCT 4 continued to be increased in vaginal infants at week 3 of life, but by week 5 of life and thereafter the PGCTs detected in cesarean and vaginal infants were comparable. Despite these trends in PGCT development, the trajectory of gut microbiome development was highly variable within and between infants and no significant difference (P = 0.125) in PGCTs was found between cesarean and vaginal infants  clustering based on weighted UniFrac, the early preterm microbiome was defined by six distinct clusters, termed PGCTs , sampled during NICU and post discharge, birth mode was not significantly associated with the alpha- or beta- diversity (both within and between patients). Furthermore, birth mode had no significant association with the relative abundance of any bacterial genera or PGCTs. A novel analysis temporally tracking OTUs within infants in consecutive samples showed vaginally delivered infants retain more OTUs from month 2 of life and post discharge, suggesting increased microbiome stability associated with vaginal delivery.Bacteroides between term and preterm infants. Bacteroides is the primary genera significantly altered between cesarean and vaginal term infants over the 1st year of life  of infants. Further study of post discharge samples is needed to determine the long-term associations of birth mode and specifically if  infants . Third,  infants .In summary, in a single NICU preterm population exposed to antibiotics postpartum, birth mode was not significantly associated with the diversity or composition of the microbiome. Vaginal infants tended to have greater stability following month 2 of life and post discharge. These findings highlight key differences between preterm and term infants. The long-term effects on host immune development from the transfer of microbes during delivery and the subsequent risk of obesity, autoimmunity, and allergy warrant further investigation.CS, NE, GT, CO, JB, and SC were responsible for the study concept and design. CS and AN were responsible for extraction and sequencing of samples. CS, EC, PL, DS, TF, and JP were involved in the data processing and analysis. CS drafted the manuscript and all authors contributed to critical revisions and approved the final manuscript.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.


Approaches for the prevention and treatment of hip osteoarthritis (OA) remain limited. There are recent data suggesting that low birth weight (LBW) and preterm birth may be risk factors for hip osteoarthritis. This has the potential to change the current paradigm of hip osteoarthritis prevention by targeting early life factors. The aim of this review was to examine the available evidence for an association of LBW and preterm birth with hip OA. The potential cost implications associated with total hip arthroplasty were also evaluated.Ovid Medline, EMBASE, and Cinahl were searched up until August 2017 using MeSH terms and key words. Methodological quality was evaluated using the National Heart Lung and Blood Institute (NHLBI) quality assessment tool. Qualitative evidence synthesis was performed to summarise the results. Bradford Hill’s criteria for causation including the temporal relationship, consistency, strength of the association, specificity, dose-response relationship, and analogy were used to assess the evidence for causation. Economic modelling was used to calculate the potential economic burden associated with LBW or preterm birth related total hip arthroplasty using Australian data from 2012 to 2015.Five studies, ranging from high to low quality, were included. Hip bone shape abnormalities examined included developmental hip dysplasia and immature hip, and hip osteoarthritis included osteophytes and total hip arthroplasty. A causal link between low birth weight or preterm birth and hip osteoarthritis was found. Of the 30,477 total hip arthroplasties performed for hip osteoarthritis in Australia in 2015, 5791 were estimated to be born preterm and 5273 with low birth weight. This equated to a potential total hip arthroplasty cost of AU$145,136,082 and AU$132,150,222 for these subgroups, respectively.Available data suggest that low birth weight and preterm birth are associated with hip bone shape abnormalities and hip osteoarthritis requiring total hip arthroplasty, with a substantial associated financial burden. Given the current lack of effective treatment and prevention strategies for hip osteoarthritis, this offers a new avenue for reducing the future burden of hip osteoarthritis.The online version of this article (10.1186/s13075-018-1627-7) contains supplementary material, which is available to authorized users. Hip osteoarthritis (OA) is a common joint disease with one in four people developing symptomatic hip OA in their lifetime . There aThere is increasing evidence for the importance of hip bone shape in the pathogenesis of hip OA , 6. For A substantial number of babies are born with LBW in developed countries. Based on data from 2015 and 2016, 6.2% of infants are born LBW < 2500 g) and 8% are born preterm (at < 37 completed weeks gestation) in Australia , which i00 g and In this study, we aimed to examine the available evidence for an association of LBW and preterm birth with hip OA, and to examine the evidence for potential causation based on the Bradford Hill criteria. Considering the range of definitions used to describe OA-related changes and outcomes, we have collectively termed ‘hip pathologies’, hip joint abnormalities, hip OA, and hip arthroplasty for hip OA as ‘Hip OA’. Given the huge burden associated with both hip OA and LBW and preterm birth, we also considered the potential cost implications associated with THA.This systematic review was conducted and reported in accordance with the Preferred Reporting Items for Systematic Review and Meta-Analysis (PRISMA) guidelines .Ovid Medline, CINAHL, and EMBASE databases were searched between January 1947 and August 2017 using MeSH terms (after exploding) and key words to identify studies examining the association between LBW or preterm birth and hip bone abnormality and hip OA. The MeSH and key terms used to define hip bone abnormality and hip OA included “hip” and “osteoarthritis” or “degenerative arthritis” or “coxarthritis” or “dysplasia” or “joint space narrowing” or “osteophytes” or “bone shape” or “bone geometry” or “neck shaft angle” or “hip deformity” or “femora acetabular impingement” or “pincer deformity” or “cam deformity”. For LBW and preterm birth, the MeSH and key terms used were “birth weight” or “low birth weight” or “preterm birth” or “prematurity” or “preterm”. Searches were limited to human studies and those published in English.Two authors (SMH and YW) independently assessed study eligibility using a three-stage determination method, reviewing the title, abstract, and then full text. Any disagreement between the two authors was resolved by consensus reviewing the criteria. Studies were only included if they assessed hip bone abnormality and hip OA including developmental dysplasia of the hip (DDH), hip instability at childhood, hip deformity, hip osteophytes, joint space narrowing, symptomatic hip OA, radiological hip OA, magnetic resonance imaging (MRI) changes of the hip, hip bone shape, hip bone geometry, hip neck-shaft angle, hip deformity, THA for hip OA, and either LBW or preterm birth or both LBW and preterm birth. Case reports, conference abstracts, and review articles were excluded, as were studies without a comparison group. The reference lists of the included articles and review articles identified were searched to identify any additional relevant studies.Two authors (SMH and YW) extracted data independently on study design, participant characteristics , definition and prevalence of hip bone abnormality and hip OA, duration of follow-up, measures of LBW or preterm birth, adjustment for confounding factors, and associations between LBW or preterm birth and hip bone abnormality and hip OA. A third author (FMC) assessed the consistency of extracted data. Although a meta-analysis was planned, significant heterogeneity across the studies  precluded the pooling of data for analysis.Two authors (INA and YW) independently assessed the risk of bias of included studies using the National Heart Lung and Blood Institute (NHLBI) quality assessment tool for observational studies . This toThe Bradford Hill criteria  were useAn economic evaluation was undertaken to estimate the costs associated with THA likely attributable to LBW or preterm birth at a national level. Trends in the national prevalence of LBW and preterm birth were examined using annual birth data published by the Australian Institute of Health and Welfare , 30. Datn = 49; Ovid Medline, n = 46; and EMBASE, n = 136), of which 69 articles were duplicates. Of the remaining 162 articles, 154 were excluded after title and abstract screening as these studies did not assess hip OA in relation to LBW or preterm birth. Full-text screening was performed for eight articles. Three were excluded as they did not include an appropriate comparison, leaving five included studies . However, the study was performed in Australia where there is a publicly funded universal health system (Medicare) and people without private health insurance have access to arthroplasty surgery under this system. The evidence of LBW and preterm birth being risk factors for abnormal hip bone shape, and hip OA was established by applying the Bradford Hill criteria of causation in two high- to fair-quality cohort studies and a few poor-quality case-control and cross-sectional studies. Thus, it is clear that further research is required to determine the influence of LBW and preterm birth in the pathological process of hip OA development. LBW could be due to prematurity (59% to 70% of low birth weight babies) , 56, intThe LBW and preterm populations were treated separately for the economic evaluation, although there will undoubtedly be some overlap between these groups. The THA costs are based on average costs for an episode of care and cannot account for individual differences in arthroplasty costs. Finally, our analysis does not include the indirect or out-of-pocket costs of THA.Despite the lack of high-quality studies in this area, our findings suggest that LBW and preterm birth are potential risk factors for hip bone shape abnormalities and hip OA requiring THA in adulthood. Based on our calculations, this could have substantial financial implications for healthcare systems. Given the current lack of effective treatment and preventive strategies for hip OA, this is an area where further research is needed to reduce the burden of hip OA. For example, the individuals born preterm or with LBW may be identified as an “at-risk group” for future end-stage hip OA; this will enable targeted monitoring and early interventions that could potentially reduce the population burden of THA in later life.Additional file 1:Table S1. Risk of bias assessment of cohort and cross-sectional studies. Table S2. Risk of bias assessment of case-control studies. (DOCX 28 kb)


Birth weight is the most important factor that affects infant and child mortality. The most common cause of low birth is malnutrition before and during the pregnancy period. The present study was conducted to explore the associated risk factors of low birth weight which will be helpful to undertake effective measures to reduce the incidence of the low birth weight babies.The study was conducted at Shaikh Khalifa Bin Zayad Al-Nayan Hospital Muzaffarabad. A sample size of 1603 live births over a period of five months were analyzed. The pregnant women were recruited in the last trimester of their pregnancy and followed up till their delivery. Information regarding maternal age, parity, income of family, gestational age, maternal occupation, degree of illiteracy, birth interval was collected. The birth weight was recorded within 24 hours of delivery. Data analysis was done using Graph Pad Prism version 6.0.Data of 1863 birth out of which 1603 were live births and among these live births, 1442 were normal birth weight babies and 161 number of low birth weight (LBW) were analyzed. The incidence of LBW in this study was (10.04%). LBW was more common in female (n=84) than in male (n=77) babies. However, this difference was statistically insignificant. Among different risk factors maternal age (p<0.05), parity (P = 0.0167), income of family (P = 0.0190), has a statically significant association with incidence of LBW. The gestational age of mother, maternal occupation, degree of illiteracy was found to affect the incidence of LBW babies, however the difference was found to be statistically insignificant for LBW. Birth interval less than three years and low hemoglobin level (P<0.0260) was found to have a significant association with LBW babies.LBW a common problem in Pakistan is an important factor for perinatal mortality and morbidity. Among different risk factors maternal age, parity, income of family, gestational age of mother, maternal occupation, degree of illiteracy, birth interval less than three years were found to be the important risk factors contributing to LBW babies born in Shaikh Khalifa Bin Zayad Al- Nayan Hospital Muzaffarabad. The babies born with under 2500g weight are term as “low weight” babies. The death rate in low weight new born is forty times compared to normal weight new born.4Many factors are known to have been associated with LBW babies such as young and old maternal age, multiple gestations, obesity, low socioeconomic status, short stature and previous preterm delivery.9In our study the objective was to identify the incidence and the common associated risk factors of low birth weight babies born in Shaikh Khalifa Bin Zayad Al-Nayan Hospital Muzaffarabad.The present study was conducted in Department of Obstetrics and Gynecology, Shaikh Khalifa Bin Zayad Al-Nayan Hospital, Muzaffarabad, Azad Jammu and Kashmir during the period of February 1, 2013 to June 30, 2013. Mother or close relatives of low birth weight babies were interviewed in labor room and obstetrics ward. The information collected included the birth weight of new born babies, birth order of babies or parity, previous still births and abortions, consanguinity, maternal age, gestational age, birth interval, blood pressure, hemoglobin tests, age at marriage, socio economic status, occupation and maternal education. Any abnormality in mothers like malaria, T.B, eclampsia, diabetes, trauma, hypothyroidism and anemia was collected. Data was collected by using a structured questionnaire. Subjects with incomplete data were excluded. Parity wise birth weight was recorded by using digital scale. At delivery, the gestation age was estimated from ultra-sound data and mother’s menstrual history. Graph Pad Prism version 6.0 was applied for the analysis.This study was approved by the ethical committee of the hospital. All procedures performed during this study and involvements of subjects were in accordance with the ethical standards of the institutional and national research committee and with the 1964 Helsinki declaration and its later amendments.Out of total 1863 recruited pregnant women had still born 70 (3.75%) had still bith, abortions accounted for 190 (10.19%) and 1603 (86.04%) were live births. Among these live births, 1442 (89.95%) were normal birth weight babies (NBW) while number of low birth weight (LBW) babies was 161 (10.04%) . In thisChi square analysis indicated that gender of the babies did not affect the incidence of LBW or NBW babies . DiffereMost patient with LBW babies (54.03%) were observed in lowest income group whereas only 3.1% were observed in high income group. Regression analysis of variance showed that this decrease in percentage of LBW babies in relation to the income of their parents was significant from zero (P = 0.0190). Proportion of LBW babies was higher (59.6%) in mothers who were house wife while this percentage was lowest 17.39% in mothers belonging to labor class. Degree of literacy of mothers also affected the incidence of LBW as large number of patients (59%) had education less than matriculation including 16.77% illiterate. The birth interval less than three years was noted in 79.50% of patients with LBW. Maternal anemia was seen in LBW patients. Maternal hemoglobin level was found to have a strong association with incidence of LBW babies. Regression analysis revealed that with an increase in hemoglobin level significantly (P<0.0260) There was decrease in the incidence of LBW babies. The number and percentage of LBW babies with different maternal factors is shown in Incidence of LBW 10.04%) observed in this study is comparable to that observed in the study carried out in in Dow Medical College & Layari General Hospital Karachi where the frequency of LBW was found as 10.6%.04% obsest parity was 71 (32.87%) which was the highest percentage in this study. This finding correlates with study done by Kayastha and Tuladhar (2007),16In present study, total NBW male and female babieswere found to be 50.34% and 49.65% respectively whereas, total LBW male and female babies were 47.82% and 52.17%. In the current study, highest percentage (48.44%) was observed in maternal age group (20-25years). This is comparable to the study from Horn, et al. (1983); Mondal (1998); Radhakrishan (2000).It was observed that in relation to maternal education 16.77% illiteracy was present in LBW mothers. Effect of rate of literacy of the mothers showed statistically significant increase on the birth weight of babies. Similar results were reported by Dhanker (2013), Joshi et al. (2010).th month resulted in higher number of LBW babies (43.47%) compared to the pregnancies that terminated at 9th month (21.73%). Similar results were also reported by Tema (2006)and Siza (2008).The pregnancies that terminated at 8 by Tema 06and Sizth month .73%. SimIn present study, the prevalence of LBW babies in anemic mothers was high 78.8% Similar results were also reported by Badshah et al.(2008).st, month of gestation, low income of family, maternal occupation, illiteracy, short birth interval. Low hemoglobin level and consanguinity were main risk factors found among LBW babies born in Shaikh Khalifa Bin Zayad Al-Nayan Hospital Muzaffarabad, Azad Jammu and Kashmir. The present study recommends that there should be counseling of mothers related to education, not to get marry at an early age, maintaining a birth interval more than three years, supplementation of iron during pregnancy and to avoid the cousin marriages.LBW a common problem in Pakistan is an important factor for perinatal mortality and morbidity. Among different risk factors the maternal age, parity 1Gulnaz Iltaf, did data collection and manuscript writing.Beenish Shahid, worked on concept, designed and supervised the study, review and final approval of manuscript.M. Ijaz Khan, did statistical analysis.


Pregnancy and birth outcomes have been found to be sensitive to meteorological variation, yet few studies explore this relationship in sub-Saharan Africa where infant mortality rates are the highest in the world. We address this research gap by examining the association between meteorological factors and birth weight in a rural population in southwestern Uganda. Our study included hospital birth records (n = 3197) from 2012 to 2015, for which we extracted meteorological exposure data for the three trimesters preceding each birth. We used linear regression, controlling for key covariates, to estimate the timing, strength, and direction of meteorological effects on birth weight. Our results indicated that precipitation during the third trimester had a positive association with birth weight, with more frequent days of precipitation associated with higher birth weight: we observed a 3.1g (95% CI: 1.0–5.3g) increase in birth weight per additional day of exposure to rainfall over 5mm. Increases in average daily temperature during the third trimester were also associated with birth weight, with an increase of 41.8g (95% CI: 0.6–82.9g) per additional degree Celsius. When the sample was stratified by season of birth, only infants born between June and November experienced a significant associated between meteorological exposures and birth weight. The association of meteorological variation with foetal growth seemed to differ by ethnicity; effect sizes of meteorological were greater among an Indigenous subset of the population, in particular for variation in temperature. Effects in all populations in this study are higher than estimates of the African continental average, highlighting the heterogeneity in the vulnerability of infant health to meteorological variation in different contexts. Our results indicate that while there is an association between meteorological variation and birth weight, the magnitude of these associations may vary across ethnic groups with differential socioeconomic resources, with implications for interventions to reduce these gradients and offset the health impacts predicted under climate change. Climate change disproportionately impacts health for those already facing the greatest burden of ill health and social inequality , 2. AdveThere has been increasing interest and research into the extent to which meteorological variation impacts foetal development –13. The We contribute to the small but emerging evidence base examining meteorological associations with birth outcomes in vulnerable populations in sub-Saharan Africa. More specifically, we evaluate the association between meteorological factors and foetal growth in a sample of births from Bwindi Community Hospital in Kanungu District, Uganda. We hypothesize that meteorological drivers will be associated with foetal growth, that these effects will be greatest in the third trimester , and thSize at birth is a function of the length of gestation and the rate of growth in utero . Birth wThe aetiologic determinants of IUGR differ from those of preterm birth . In the Our study sample consists of 3691 women who gave birth at Bwindi Community Hospital in Kanungu District, Uganda, between June 2012 and June 2015. Kanungu District is situated in southwestern Uganda, near the borders of the Democratic Republic of the Congo and Rwanda . The stuResearch has documented high vulnerability to the health impacts of climate change within both the Batwa and Bakiga populations, and both groups have identified malaria, food insecurity, and gastrointestinal illnesses as climate-sensitive health concerns , 29. LacBwindi Community Hospital (BCH) was established in 2003 as an outreach clinic for the Batwa, but has expanded into a full inpatient hospital serving 100 000 people across three sub-counties in Kanungu district . The hosThe climate in Kanungu includes two rainy seasons: the ‘short rains’ from October to December and the ‘long rains’ from March to May . AverageBirth outcomes were ascertained from the hospital records that were completed by nurses during labour and after birth. They include information on the mother’s medical history, intranatal measures and interventions, and assessments of the baby at birth. Birth weight in grams is our primary outcome. In the available birth record data from Bwindi Community Hospital (2012–2015), there were 3691 births, 3343 of which were singleton births. Of these, there were 3197 complete records with sufficient information to estimate gestational age.Gestational age (GA) was estimated for all singleton birth records (n = 3197) to determine the exposure windows of the meteorological variables and to include as a key control variable in our ‘base’ regression models for the primary outcome analysis. In the case of 591 observations (17.9%), gestational age was recorded based on ultrasound dating, the most reliable measurement standard for GA. The date of the ultrasounds were not collected, but we note that ultrasounds conducted later in pregnancy have a greater margin of error than those conducted within the first trimester, and that there is variation in the timing of initial presentation of pregnant women to the hospital or antenatal clinic. Where GA based on ultrasound was not available, last menstrual period  and adjusted fundal height  were used. Fundal height measures  upon presentation for delivery were compared with ultrasound expected delivery dates and adjusted to control for fundal height underestimation of gestation length by an average of two weeks. If the gestational age based on LMP exceeded 330 days , fundal Meteorological predictor selection was based on the literature and variation in weather in the study area. We selected the number of days of precipitation ) above 5mm during the exposure window (described below) as our meteorological exposure for the effects of precipitation in our models. Given the limited range of temperature fluctuations in the region, the mean of daily temperature (°C) during each exposure period was selected to examine the effects of temperature on the outcome variables. Birth season was included in all models as a fixed effect to control for any unaccounted effects of seasonality .Meteorological data for rainfall and temperature were extracted and matched to each birth for multiple exposure periods reflecting the entire pregnancy period and each trimester. Daily rainfall data were estimated based on satellite observations using the Rainfall Estimator, version 2.0 (RFE2) algorithm , 51. In We controlled for variables known to influence birth weight . We inclWe included ethnicity to compare Indigenous vs. non-Indigenous birth outcomes in the population, which in this context serves as a partial proxy for socio-economic status (see Section 2.1). We hypothesized that ethnicity would act not only as a control variable, but also an effect modifier for meteorological effects on birth weight. This hypothesis is consistent with climate change adaptation literature, which theorizes that climatic and/or meteorological effects will manifest through, and be modified by, existing gradients in health , 6, 7. OOur analytical sample included live singleton births with information on birth weight, gestational age, and all known covariates (n = 3197). Linear regression was used to estimate mean differences in birth weight associated with meteorological exposures. We constructed a ‘base’ model that includes infant sex and gestational age to examine sex- and gestational age-specific birth weight differences by exposure and potential confounders. Gestational age was included as a categorical control variable due to the potential relatively high measurement errors inherent to LMP- and fundal height-based measurements and a non-linear relation to birth weight suggested by the lowess-smoothed scatterplot.We then built models that included all control variables and meteorological exposures, constructing four models: a model with meteorological exposures and all control variables for each of the three trimesters, and a fourth model with meteorological exposures and all control variables for the entire pregnancy period. We stratified the sample by ethnicity and tested the same models in each group to investigate ethnicity as an effect modifier. In a separate analysis, we stratified the sample by season of birth and tested the models for each exposure period.We conducted multiple sensitivity analyses to examine robustness of associations: (1) re-analyzing the sample after excluding preterm births (<37 weeks), (2) re-analyzing the sample after excluding all cases for which GA was not determined by ultrasound, (3) examining associations using different thresholds for temperature  and precipitation , (4) testing for interaction effects between meteorological exposures and infant sex, and (5) testing models with month of conception instead of season of birth to evaluate any non-random seasonal fertility selection. Further sensitivity analyses evaluated LBW as an outcome using logistic regression. This analysis allowed us to examine the lower 5–10% of birth weights separately, since as Grace, Davenport  state, tTo assess model fit, we evaluated collinearity by examining the Pearson correlation coefficient matrix for all predictor variables and control variables. We conducted post-estimation tests to assess heteroscedasticity (Breysch-Pagan test) and the normality of the residuals (Q-Q plots and Shapiro-Wilk tests) in our models. We checked that the outcome variable was normally distributed . We also evaluated the linearity of the meteorological exposure variables and other control variables using scatterplots, as well as lowess smoothing to visually assess trends in these plots. Data were analysed using Stata v.13 .This study was approved by the McGill University and the University of Guelph Research Ethics Boards (REB File #461–0414). Additionally, the research team has a Memorandum of Understanding with, and received approval for this study from, Bwindi Community Hospital (BCH). The study design conforms to the Canadian Tri-council Policies and follows the requirements of the Ethical Conduct of Research Involving Human Subjects; it is also in compliance with Ugandan laws and regulations for foreign researchers. As per McGill, Tri-Council, and BCH ethical research policies, informed consent by individual patients specific to this study was not required. This study included retrospective analysis of de-identified hospital records, and consent to use the data for hospital and hospital-approved research analysis is provided by the patient at the time of admission to the hospital or maternity ward.The average age of mothers in the study was 24.66 (SD: 5.76) years old, with a mean of 2.72 previous pregnancies (SD: 2.09) and a mean parity of 1.63 (SD: 1.86) births per woman . The majBirth weight increased significantly with maternal age, among married mothers, in babies born via C-section, and with increasing parity after controlling for gestation age and infant sex . The resIn models testing crude associations of exposure variables with birth weight  , exposurMultivariable models tested the effects of meteorological predictors on birth weight adjusting for all covariates in Strikingly, given the small sample size of the Batwa subset, mean temperature exposure during the entire pregnancy was positively associated with birth weight in the Indigenous (Batwa) subset . Each deIn addition to this association observed between birth weight and temperature exposure during the third trimester, sensitivity analyses of models for the Batwa subset with an adjusted precipitation variable with a lower threshold  and of models restricted to term births (≥37 weeks) indicated that there may also be a significant association between mean temperature exposure incurred throughout pregnancy and birth weight.The trends we observed in Results of our sensitivity analyses of subsets restricted to term births and cases with ultrasound GA measures showed consistent results in both direction and statistical significance of associations . Models We also tested for interaction between sex and meteorological values and found no significant interactions. Lastly, we included dummy variables for month in the full models and observed no changes in the patterns of association.To our knowledge, this is one of the first studies to examine the effects of specific perinatal meteorological exposures on birth weight in a vulnerable sub-Saharan Africa population, and the first to consider effect modification by existing social gradients in health in a specific regional setting. We find comparable results to Grace, Davenport  with resSeason of birth did not emerge in the initial models as a significant determinant of birth weight. However, when we stratified the sample by season of birth, the associations between meteorological exposures and birth weight were inconsistent across babies born at different times of year. Indeed, only those born between June and August showed significant associations between in utero temperature and precipitation exposures; those born between September and November showed a significant exposure only between third trimester average temperature exposure and birth weight. This heightened sensitivity to meteorological exposures for July through August births (the driest and hottest season) could be linked to harvest cycles in the region—women who give birth in this season may benefit more from increased temperatures and increased rainfall in their third trimester as these conditions could increase the harvest yield in a time of year when food is most scarce. The third trimester is a crucial period for increasing birth weight, leading us to conjecture that the reverse trends observed for second trimester exposures may be residual effects related to a preferencing of the most optimal conditions during the last trimester.The explanation of the positive effect of temperature on birth weight in season 4 (September to November) is less obvious, and may be related to the effects of temperature on local harvest cycles  or potentially to more direct physiological effects of ambient temperature on foetal development , 58. VarIn the non-Indigenous subset, it was primarily weather exposures incurred in the third trimester that were associated with birth weight. However, the effect of mean temperature exposure appeared to be important throughout pregnancy in several models examining only Indigenous Batwa cases. We found some variation in the associations between meteorological exposures and birth weight among Batwa compared to the non-Indigenous population. This may reflect modification of meteorological impacts on birth outcomes by ethnicity among mothers living in the same region and with access to similar health services.The differential impacts of temperature on birth weight between the Indigenous and non-Indigenous subsets may reflect differences in agricultural cycles between the two populations. Patterson et al. (Under review) found, for example, that the Batwa seasons of harvest and famine begin earlier than neighbouring non-Indigenous populations in Kanungu District due to high impoverishment and limited historic experience with agriculture. Thus, the direct physiological effects of ambient temperature on birth weight may supersede those of variation in food availability according to seasonal precipitation in this population. This result is consistent with findings from Grace, Davenport  that shoOne of the chief limitations to this study is its reliance on hospital births, a sample which is likely not representative of all births in the area and thus limits the generalizability of our findings. In 2013–2014, for example, only 1264 (42%) of the estimated 2958 deliveries in the catchment area occurred in the hospital . This spOur results highlight heterogeneity in the vulnerability of infant health to meteorological variation in different contexts, with implications for climate change impacts and adaptation research. The health effects of climate change will manifest along existing social gradients, with gendered impacts on nutrition highlighted as an area of particular concern since women and girls’ nutrition is often deprioritized relative to their male relatives . Our resPregnancy results in increased energy consumption and increased nutritional needs , renderiThe results of this study are of particular relevance to maternal health services at Bwindi Community Hospital, which aims to decrease maternal and child mortality rates by 25% by 2019 . As clim


Bacterial vaginosis (BV) during pregnancy is a well-established risk factor for preterm birth and other preterm pregnancy complications. Little is known about adverse neonatal outcomes associated with BV exposure in full-term births, nor its influence on adverse outcomes independent of its effect on gestational age. The purpose of this study was to examine the relationship between BV during pregnancy and adverse neonatal outcomes among full-term and preterm infants.We conducted a retrospective cohort study of Washington State mother/infant pairs from 2003-2013, stratified by full-term (primary outcomes) and preterm births (secondary outcomes). BV-exposed and unexposed women were frequency-matched based on year of delivery. BV exposure and adverse outcomes  were identified using birth certificates, ICD-9 codes from linked hospital records, and death certificates. Associations between BV exposure and outcomes were assessed using multivariable Poisson regression, adjusted for maternal demographics, gestational age, and other pregnancy complications, including infections.A total of 12,340 mother/infant pairs were included: 2,468 BV-exposed  and 9,872 BV unexposed . Among full-term infants, BV-exposed mothers were younger, more likely to be Black or Hispanic, more likely to have had a sexually transmitted infection, and less likely to have a college degree than unexposed mothers. Term BV exposed infants were more likely to have meconium at delivery. Following adjustment, BV was associated with an increased risk of assisted ventilation/respiratory distress at birth , NICU admission , and neonatal sepsis  among full-term infants. These associations were independent of the presence of chorioamnionitis or meconium. Among preterm infants, BV-exposure was associated with an increased risk for NICU admissions only .BV exposure during pregnancy is associated with adverse neonatal outcomes even among infants born full-term. These findings amongst full-term infants are novel, and highlight neonatal implications of BV in pregnancy independent of BV’s effect on preterm birth. Lactobacillus and overgrowth of anaerobic bacteria  of newborn) and 771.83 (bacteremia of newborn) and by birth certificate checkbox. Because this study primarily used birth certificate data, fetal deaths before 20 weeks gestation (where no birth or death certificate would have been issued) were not captured. Fetal mortality at 20 weeks or greater gestation was identified from birth hospitalization records, ICD-9 code V27.1, and linked death certificates. Infant mortality is defined as death within one year of birth, and was identified via linked birth and death certificates.N = 12,317); when this variable was missing, we estimated gestational age using the date of the mother’s last menstrual period (N = 23). We established a priori that maternal age and maternal race  would be included in the models [We used Poisson regression to calculate risk ratios with 95 % confidence intervals for all outcomes among sub-cohorts of full-term infants (≥37 weeks gestation) and preterm infants (<37 weeks gestation), as well as the complete cohort. Gestational age was established from the birth certificate, primarily from ultrasound dating in the first or second trimester  [In addition, we considered the following as covariates: maternal education ; delivery method ; prenatal smoking (yes/no); nulliparity (yes/no); birth weight <2500 grams (yes/no); small for gestational age among live born infants  ; sexuallAll analyses were conducted using STATA 13 software .N = 11,354) and pre-term (N = 975) sub-cohort analyses but were included for analyses of the full cohort. There were 2,198 term and 267 preterm mother/infant pairs identified that were BV exposed.We included and analyzed 12,340 mother/infant pairs. Gestational age was missing for 11 infants; these participants were excluded from the full-term  . We confWe did not find an increased risk for term perinatal mortality associated with BV and, to our knowledge, this has not been studied previously. BV has not been associated with perinatal mortality in previous meta-analyses of primarily preterm cohorts . PerinatOur study has strengths and limitations. It was large, population-based, and rich in covariates, with little missing data. While birth certificate sensitivity is poor for some variables , 25, we BV during pregnancy appears to increase the risk for adverse neonatal outcomes through mechanisms independent of BV’s effect on timing of delivery. These findings may have significant clinical implications if confirmed by other studies, but several questions remain, including the importance of timing of the exposure and the mechanism for these associations. While treatment of BV during pregnancy has not succeeded in reducing preterm birth , 27, tre


To establish the contribution of maternal, fetal and intrapartum factors to the risk of incidence of obstetric anal sphincter injuries (OASIS) and assess the feasibility of an OASIS risk prediction model based on variables available to clinicians prior to birth.This was a population-based, retrospective cohort study using single-site data from the birth database of Aarhus University Hospital, Denmark. The participants were all women who had a singleton vaginal birth during the period 1989 to 2006. Univariate and multivariate logistic regression analyses were performed using multiple imputations for missing data and internally validated using bootstrap methods. The main outcome measures were the contributions of maternal, fetal and intrapartum events to the incidence of OASIS.A total of 71,469 women met the inclusion criteria, of whom 1,754 (2.45 %) sustained OASIS. In the multivariate analysis of variables known prior to birth, maternal age 20 – 30 years  and ≥30 years , occipitoposterior fetal position , induction/augmentation of labour , and suspected macrosomia  were independent significant predictors of OASIS, with increasing parity conferring a significant protective effect. The ‘prebirth variable’ model showed a 95 % sensitivity and a 24 % specificity in predicting OASIS with 1 % probability, and a 3 % sensitivity and a 99 % specificity in predicting OASIS with a 10 % probability.Our model identified several significant OASIS risk factors that are known prior to actual birth. The prognostic model shows potential for ruling out OASIS , but is not useful for ruling in the event.The online version of this article (doi:10.1007/s00192-016-3125-2) contains supplementary material, which is available to authorized users. Obstetric anal sphincter injuries (OASIS) are serious complications of vaginal birth with a reported incidence throughout Europe and globally from 1 % to 10 % . Recent With the wealth of available population-based databases, prediction models are being developed to assist in clinical decision making . SeveralThe aims of this study were twofold. The primary aim was to explore both the burden of the contribution of each variable and the combined effects of multiple factors on the risk of OASIS in singleton pregnancies. The secondary aim was to develop an OASIS prognostic model using variables known only prior to birth and to evaluate its feasibility for accurately predicting the individual risk of OASIS in singleton pregnancies to provide women and clinicians with information for making informed decisions about intrapartum management and mode of birth.This was a retrospective population-based cohort study using the birth cohort database of Aarhus University Hospital, Denmark. All women who had singleton pregnancies and a vaginal birth during the period 1989 – 2006 were included. For the purposes of our analysis the classification of OASIS in the current RCOG Green-top Guideline was used , and allt tests for continuous data, and the Mann-Whitney U test was used for continuous skewed variables and the chi-squared test for categorical variables.The study sample was first summarized by comparing women who sustained and women who did not sustain OASIS, reporting means and standard deviations . The two groups were compared using A comprehensive stepwise procedure was used to select independent factors predicting OASIS. This included the evaluation of two logistic regression models reporting odds ratios (OR) and 95 % confidence intervals (CI). The first model  was developed using all independent predictive factors available . This allowed consideration of all factors available in the database. However, as some of these factors are only known after birth they may not necessarily be useful as part of a clinical prediction model. Consequently, a second logistic regression model (OSIRIS – Prebirth Variables model), reporting OR and 95 % CI and including predictive factors that could only be determined or planned before the actual birth of the baby, was developed. These variables included maternal age, gestational age, BMI, parity, induction/augmentation of labour, fetal position, and the presence of fetal macrosomia. In our database, the actual birth weight was documented for each birth, but no assumption as to the presence of macrosomia was recorded prior to delivery. Although ultrasonographic biometric indices and the accuracy of estimated fetal weight remain low, they are improving ; therefoP value for addition set at 0.05 and for removal set at 0.1) complemented with a sensitivity check involving the implementation of a backwards stepwise selection procedure. Given the large size of the dataset and the likelihood of statistical significance being obtained, a further criterion for selecting a predictor was that it should obtain a prespecified clinical importance .Stepwise selection methods were used to identify independent predictive variables. This was done independently for each set of the two prediction models developed. This consisted of a forward stepwise selection procedure (with a z scores) within the stepwise regression models, to allow selection of the most important variables.Multiple imputations using chained equations with predictive mean matching was used as all predictors had at least some data missing. Ten multiple imputations were used and continuous variables were analysed as linear predictors within the multiple imputations but included as categorical or continuous covariates (The discriminatory abilities of the models produced (to distinguish between those who will and those who will not have an adverse outcome) were evaluated using the C statistic   in parity, fetal malposition, use of induction/augmentation, episiotomy, instrumental delivery, use of epidural anaesthesia, birth weight ≥4,000 g, duration of pushing phase, head circumference, gestational age and actual birth weight.A total of 71,469 women had a vaginal birth in the unit during the study period, and of these 1,754 (2.45 %) sustained OASIS, an incidence slightly lower than the overall UK rate . Table 1In the multivariate analysis of all significant factors (OASIS – All Variables model) induction/augmentation of labour , pushing for more than 30 min with the effect increasing with the duration of the active second stage and the largest effect seen in those pushing for 120 min or more , head circumference ≥37 cm , and actual fetal weight  remained significant risk factors for sustaining OASIS, whereas mediolateral episiotomy had a protective effect  maternal age 20 – 30 years vs. <20 years  and ≥30 years vs. <20 years , occipitoposterior fetal position , induction/augmentation of labour , suspected macrosomia based on the assumption that babies with a birth weight of ≥4,000 g would have been accurately suspected antenatally  all remained as significant predictors of OASIS  for the OSIRIS – All Variables model, and 0.71 (95 % CI 0.70 – 0.72) for the OSIRIS – Prebirth Variables model. The latter model also showed good calibration (not shown) throughout the risk range. The internally validated bootstrapped multiple imputation results Table .The discriminatory ability of the OSIRIS – Prebirth Variables model was further evaluated by determining the sensitivity and specificity for various probability cut-off values Table . Taking Our study identified several factors that are significant predictors of OASIS, but several of these changed following the multivariate analyses for both models. The factor with the highest contribution to the risk of OASIS when all significant variables were taken into account (OSIRIS – All Variables model) was the duration of the pushing phase, a result not consistently supported by other reports , 18, pos Some variables were identified as significant in both models, for example the use of oxytocin for induction or augmentation of labour. A similar association has previously been reported by Poen et al. . TherefoThe main strength of our study lies in the large database used in the analysis that provided a wide variety of factors enabling the development of a clinically meaningful model that excluded variables unknown until after delivery. However, we recognize that the study had several limitations. Because of the retrospective nature of the data, it was not possible to assess the accuracy of perineal trauma assessment, to determine the risk of occult tears, to determine if oxytocin had been used for induction or augmentation, to obtain information about the previous history of OASIS, or to determine if intrapartum interventions known to modify the risk of OASIS, such as manual perineal support , 28 or wAlthough we were able to prove the feasibility of a statistically robust prognostic model for predicting the individual risk of OASIS using demographic and obstetric factors known prior to birth, we were not able to demonstrate its projected usefulness in a clinical setting to rule in OASIS. Indeed, the OSIRIS Prebirth Variables model would categorize both women who have a few risk factors and those who have all possible risk factors as high risk. Undoubtedly, this high false-positive rate could lead to undue anxiety and potentially and unnecessary interventions. Nevertheless, this model seems to be good at ruling out OASIS; however, the clinical utility of this feature requires further investigation.Table S1(DOC 52 kb)Below is the link to the electronic supplementary material.