APOE-ε4 genetic status, and level of vascular risk.This study examined the association of lifetime experiences, measured by a cognitive reserve (CR) composite score composed of years of education, literacy, and vocabulary measures, to level and rate of change in white matter microstructure, as assessed by diffusion tensor imaging (DTI) measures. We also examined whether the relationship between the proxy CR composite score and white matter microstructure was modified by participant age, n = 166 cognitively normal, n = 26 mild cognitive impairment) older adults [mean age = 70.17 (SD = 8.5) years] from the BIOCARD study underwent longitudinal DTI . White matter microstructure was quantified by fractional anisotropy (FA) and radial diffusivity (RD) values in global white matter tracts and medial temporal lobe (MTL) white matter tracts.A sample of 192 non-demented (APOE-ε4 genotype or level of vascular risk.Using longitudinal linear mixed effect models, we found that FA decreased over time and RD increased over time in both the global and MTL DTI composites, but the rate of change in these DTI measures was not related to level of CR. However, there were significant interactions between the CR composite score and age for global RD in the full sample, and for global FA, global RD, and MTL RD among those with normal cognition. These interactions indicated that among participants with a lower baseline age, higher CR composite scores were associated with higher FA and lower RD values, while among participants with higher age at baseline, higher CR composite scores were associated with lower FA and higher RD values. Furthermore, these relationships were not modified by The association between level of CR and DTI measures differs by age, suggesting a possible neuroprotective effect of CR among late middle-aged adults that shifts to a compensatory effect among older adults. Cognitive reserve (CR) is commonly defined as an attribute of the brain that provides for better-than-expected cognitive performance in the presence of age- or disease-related brain changes  affect brain white matter microstructural properties over time, as assessed by diffusion tensor imaging (DTI). DTI is a magnetic resonance imaging (MRI) technique for measuring the diffusion of water in tissue. Higher fractional anisotropy (FA), a measure of directional diffusion, and lower radial or mean diffusion  in white matter tracts are commonly considered to represent greater white matter structural “integrity” , though Relatively few studies have investigated how cognitive reserve proxies, such as education, may affect white matter microstructure among middle-aged and older individuals with normal cognition and those with mild cognitive impairment (MCI). To our knowledge, these studies have all been cross-sectional in nature and results have been mixed. For instance, APOE-ε4 genetic status  for the discrepancy in results across prior studies remain unclear, but could be related to differences in white matter tracts examined and participant characteristics, including age, proportion of individuals at genetic risk for Alzheimer’s disease (AD), level of vascular risk, and cognitive status . One goal of the current study, therefore, was to systematically investigate whether the relationship between a proxy measure of CR and white matter microstructural integrity is modified by participant age, Participants included in the current study are part of the ongoing prospective, longitudinal BIOCARD study, which was initiated at the National Institutes of Health (NIH) in 1995. The BIOCARD study has the overarching goal of identifying variables among individuals with normal cognition that could predict the onset of mild to moderate symptoms of AD. After providing written informed consent, 349 cognitively normal, primarily middle-aged participants were enrolled in the study. Approximately 75% of the participants had a family history of AD dementia. While the study was at the NIH, participants completed annual clinical and cognitive assessments, as well as biomarker assessments approximately every other year. The study was stopped in 2005 for administrative reasons and resumed at Johns Hopkins University (JHU) in 2009, upon approval by the JHU Institutional Review Board. Annual blood draws and clinical and cognitive evaluations were reinitiated in 2009. The biennial collection of MRI scans (including DTI) was resumed in 2015. Additional details on the resumption of other biomarker assessments are described in This study reports data from 192 non-demented participants with DTI data collected between 2015 and 2020. For the purposes of these analyses, “baseline” was defined as the first available DTI scan. Of note, the MRI scans collected at the NIH did not include DTI sequences, thus the first possible DTI scan in this cohort was acquired in 2015.N = 29) were included in the group of cognitively normal participants, but results were comparable when they were excluded from analysis (data not shown). All clinical diagnoses were made independently from knowledge of any biomarker measures.The annual visits at JHU include a neuropsychological battery, clinical assessments, and a semi-structured interview based on the Clinical Dementia Rating Scale CDR,  and deme2, 170 slices with 1 × 1 × 1.2 mm3 voxels, and scan duration = 5 min 59 s). Diffusion-weighted images were acquired from a spin echo sequence . DTI images from all participants were collected from the same scanner and processed with the same software and parameters.MRI scans were acquired on a 3 T Phillips Achieva scanner . The multi-modal protocol included magnetization-prepared rapid gradient echo (MPRAGE) scans used for anatomical reference and image registration  was operationalized by a composite score that included three measures commonly used as proxies of CR, hypothesized to reflect lifetime cognitive experiences: (1) years of education; (2) scores from the National Adult Reading Test NART; ; and (3)2), and current smoking .Vascular risk factors were established through self-report during a medical history interview and available medical records. A composite vascular risk score was calculated using a previously validated approach  that sumAPOE-ε4 carriers = 1, non-carriers = 0).APOE genotypes were determined by restriction endonuclease digestion of PCR-amplified genomic DNA . Genotypes were coded dichotomously . Estimates and p-values are reported, and p-values < 0.05 were considered significant. All analyses were run in R, version 3.5.0.Sensitivity analyses examined whether the patterns of results changed when individuals with MCI were excluded. Separate models were not run for the participants with MCI given the small sample size . Additionally, the mean follow-up time between participants’ first and last DTI scan was less for participants with MCI. Of note, higher CR composite scores were correlated with higher MMSE scores across the non-demented participants, partialling out age and sex, , and in the cognitively normal subgroup .Baseline descriptive characteristics of all participants included in the analyses, stratified by diagnosis, are reported in n = 192) and the subset with normal cognition (n = 166), there were significant effects of time for all DTI measures, indicated by decreases in FA values and increases in RD values over time, in both the global and MTL DTI composites . However, there were no significant CR × time interactions in either group , indicating that the rate of change in DTI measures over time does not differ by level of CR. Additionally, in reduced models that excluded the non-significant CR × time interaction terms, there were no main effects of CR on global or MTL FA or RD in the full sample . Among the subset with normal cognition, there was a significant main effect of CR on MTL RD only , indicating that higher CR composite scores were associated with lower RD in the MTL tracts.In both the full sample (p < 0.001), and female sex was associated with lower global and MTL RD values (both p < 0.04). Furthermore, an MCI diagnosis was associated with lower global and MTL FA values (both p < 0.04), as well as higher global and MTL RD values (both p < 0.03). In addition, the rates of change in global and MTL FA values, and in global RD values, over time were greater among older participants, as indicated by significant age × time interactions (both p < 0.05), with a similar trend for MTL RD values (p = 0.051). The patterns of covariate effects were similar for the subset with normal cognition, except that the age × time interactions were only significant for the global RD values (p = 0.009), whereas these interactions were trending for the global and MTL FA values and MTL RD values .For the model covariates (from the reduced models that excluded the non-significant interaction terms), in the full sample, higher age at baseline was associated with lower FA and higher RD values in the global and MTL DTI composites  and the subset with normal cognition (n = 166), there were no significant CR × age × time interactions for any of the DTI measures , indicating that the associations between the CR composite score and rates of change in the DTI measures over time do not differ by baseline age. The results for the reduced models, which excluded the non-significant CR × age × time interaction terms, are shown in In both the full sample , p = 0.013), global RD , p < 0.001), and MTL RD , p = 0.05) remained significant.When the models were restricted to the p > 0.10, data not shown).In exploratory analyses, there were similar CR × age interactions for the majority of the individual white matter tracts included in the global composite see , suggestn = 192) and the subset with normal cognition (n = 166), there were no significant CR × APOE-ε4 × time interactions for any of the DTI measures ; similarly, in the reduced models that excluded these non-significant 3-way interaction terms, there were no significant CR × APOE-ε4 interactions . This indicates that the associations between the CR composite score and level of, and rate of change in the DTI measures over time, did not differ by APOE-ε4 genotype status. Additionally, APOE-ε4 genotype status was unrelated to the levels of , and rate of change over time in , the DTI measures.In both the full sample (n = 192) and the subset with normal cognition (n = 166), there were no significant CR × vascular risk score × time interactions for any of the DTI measures ; similarly, in the reduced models that excluded these non-significant 3-way interaction terms, there were no significant CR × vascular risk score interactions . This indicates that the associations between the CR composite score and level of, and rate of change in the DTI measures over time, did not differ by level of vascular risk. Additionally, the vascular risk score was unrelated to the levels of , and rate of change in , the DTI measures.In both the full sample  found that greater occupational complexity in young to middle-adulthood was associated with higher FA globally and in all four lobes 10–15 years later, when participants were in middle age . For example, a large cross-sectional study  studies among cognitively normal older adults have linked higher educational attainment , as wellExploratory analyses further suggested that the interaction between age and CR on white matter microstructure was evident in the majority of the individual tracts included in the composite measures. Conceptually, it seems reasonable that fairly broad measures of lifetime experience  would have global effects on the brain, as opposed to highly localized effects. However, future studies are needed to replicate this finding.APOE-ε4 genotype. Additionally, APOE-ε4 genotype was unrelated to the level of, or rate of change in, global or MTL white matter microstructure, which is consistent with prior studies among cognitively normal older adults  and poorer white matter integrity, as indexed by lower FA among middle-aged and older adults e.g., . These nBDNF Val66Met polymorphism interacts with CR to enhance executive function  as one possible mechanism. For example, prior studies in rodent models have demonstrated that BDNF is upregulated in brain regions including the hippocampus among animals exposed to enriched environments that provide more cognitive and physical stimulation . BDNF isfunction . Moreovefunction . Althougfunction . Future This study has limitations. First, DTI provides only an indirect measure of white matter integrity . Thus, twww.biocard-se.org).Anonymized data used in the analyses presented in this report are available on request from qualified investigators .MA is an advisor to Eli Lily.The remaining 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.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. 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Existing atmospheric methods for separating ffCO2 from large natural carbon dioxide variations are constrained by sampling limitations, so that estimates of regional changes in ffCO2 emissions, such as those occurring in response to coronavirus disease 2019 (COVID-19) lockdowns, rely on indirect activity data. We present a method for quantifying regional signals of ffCO2 based on continuous atmospheric measurements of oxygen and carbon dioxide combined into the tracer “atmospheric potential oxygen” (APO). We detect and quantify ffCO2 reductions during 2020–2021 caused by the two U.K. COVID-19 lockdowns individually using APO data from Weybourne Atmospheric Observatory in the United Kingdom and a machine learning algorithm. Our APO-based assessment has near–real-time potential and provides high-frequency information that is in good agreement with the spread of ffCO2 emissions reductions from three independent lower-frequency U.K. estimates.It is not currently possible to quantify regional-scale fossil fuel carbon dioxide (ffCO 2 reductions from COVID-19 U.K. lockdowns in 2020–2021.“Atmospheric potential oxygen” is used to quantify fossil fuel CO In East China, a study using satellite CO2 retrievals was also unsuccessful  were used to infer COVID-19–related ffCO2 reductions, mostly from wintertime data (when biospheric activity is suppressed), and by assuming that biospheric-related variability in CO2:CH4 ratios was not different in 2020 compared to previous years  using satellite-based nitrogen dioxide (NO2) observations and CO2-to-NOx emission ratios from a bottom-up estimate to formulate proxy ffCO2 observations  and a high-resolution bottom-up ffCO2 emissions prior within an atmospheric inversion framework. The authors found a 30% reduction in ffCO2 emissions during a 6-week period of the city’s “shelter-in-place” order, compared to the previous 6-week period  the availability of high-resolution emissions or emission ratio information from bottom-up inventories, and (iii) the selection of urban-based measurement locations, where ffCO2 signals are comparatively larger [typically from 0 to 30 parts per million (ppm)] than signals at non-urban sites .At the urban scale, where the contribution of ffCO2 can also be separated with ground-based atmospheric measurements, using radiocarbon data (14CO2), carbon monoxide data (CO), or a combination of both , which are moreover expensive and laborious to analyze  using atmospheric potential oxygen (APO) data. We demonstrate the potential of APO as a ffCO2 tracer by detecting and quantifying COVID-19 ffCO2 reductions in the atmosphere associated with the first two waves of the pandemic in the United Kingdom, using continuous data from the Weybourne Atmospheric Observatory (WAO) in the United Kingdom and a machine learning algorithm. The APO-based assessment we present here separates biospheric and anthropogenic signals in atmospheric CO2 with high frequency  and in near real time, which is an important first step toward robust quantification of absolute ffCO2 emissions using atmospheric data. Our approach does not quantify absolute emissions, but, with the use of machine learning, we are able to quantify relative changes in emissions using APO data, which represents a major achievement in top-down observation-based ffCO2 emissions quantification efforts. Using a combined APO and machine learning approach, we have detected a local 1.6-ppm reduction in daily-mean ffCO2 observed at WAO during March to July 2020 compared to the nonpandemic “counterfactual scenario” , and a 1.3-ppm daily-mean reduction during November 2020 to January 2021. These two U.K. lockdown periods are separated by a period of recovery, from August to October 2020, characterized by little reduction in ffCO2. Our APO-based estimate is in good agreement with the spread of ffCO2 reductions determined from three independent bottom-up emissions estimates for the United Kingdom.Here, we present a new ground-based measurement approach for quantifying the regional ffCO2) and CO2 observations (APO = O2 + 1.1 × CO2)  of APO:CO2 for fossil fuel emissions, derived from an emissions database product  because O2 is not a trace gas and its mole fraction is therefore affected by small changes in other gases, such as CO2 (2 and CO2 fluxes (BL) thus isolates short-term anomalies in APO, from which ffCO2[APO] is determined. We convert the units of APO-APOBL from per meg to parts per million equivalent by dividing by 4.77 . Air masses arriving at the site from the North Sea and the Arctic Ocean are generally associated with lower ffCO2 compared to air masses arriving from the direction of southern England, the Midlands, or from the European continent. A cumulative plot of ffCO2 for each year during 2011–2020, shown in 2 observed during 2020 compared to previous years because the dominating influences of wind direction and atmospheric transport on the ffCO2 signals at WAO have not been accounted for.2[APO] associated with the U.K. COVID-19 lockdown restrictions, using an 11-year continuous APO dataset from WAO and a random forest machine learning algorithm  prediction during periods when COVID-19 restrictions were in place.We have applied the APO method to detect and quantify the reduction in ffCOlgorithm . A suite2 detected at WAO is highlighted using the cumulative signal, shown in 2[APO] relative to the counterfactual (nonpandemic) prediction of 1.6 ppm from 20 March to 31 July 2020, coinciding with the first period of U.K. lockdown, a recovery period during August to October 2020 during which U.K. lockdown restrictions were eased and ffCO2[APO] increased slightly by 0.2 ppm, and a second sustained decrease in daily-mean ffCO2[APO] of 1.3 ppm from early November 2020 until the end of January 2021, during which a national lockdown was reintroduced. We deem these sustained signals to be caused by reductions in ffCO2 emissions within the footprint of WAO, where the term “footprint” denotes the sensitivity of measurements at WAO to emissions located upwind of the site location .The COVID-19 influence on ffCO2 change associated with COVID-19 emissions reductions [which is ~0.3 ppm for an 8% reduction in 2020 annual emissions according to another study  from ffCO2[APO], compared to a 17% reduction from the U.K. BEIS  national inventory. Estimates based on proxy data give an 8% reduction from Carbon Monitor . This assumption relies on our ffCO2[APO] error remaining the same during both the training and predictive steps. For the APO measurement data and APOBL, there is no indication in the diagnostic data that this is not the case: Measurement performance during 2020–2021 was similar to previous years, and there is no reason why the statistical baseline fitting routine should be less accurate during this time period than previously nor any evidence in the data to suggest that this is the case.For our COVID-19 analysis, while the ffCORAPO, a shift in this value associated with COVID-19 lockdowns would potentially bias our results because we used a mean RAPO value of 0.37 throughout the whole 2010–2021 period; however, it is unlikely that any COVID-19–related RAPO shift occurred, since bottom-up estimates have shown that the reduction in ffCO2 emissions is mostly in aviation and surface transport , which shows a change in RAPO during the COVID-19 lockdown periods of only +0.01. During 2011–2019, RAPO does not change substantially except during 2012–2015, when the value changes by −0.05 because of reduced coal emissions. RAPO returns to its original 2011 value in 2016 due to increases in gas usage, which counteracts the previous influence of coal on RAPO. Thus, there is no requirement to account for ffCO2[APO] uncertainties in our COVID-19 analysis, although these do exist  and may need to be considered carefully in potential follow-up studies.For 2 but generally performs well with a relatively small bias , higher atmospheric pressure and lower wind speeds (during more stable conditions), higher radon-222 activity (more influence from ground-based sources and/or lower planetary boundary layer height), and when clustered trajectories from the Hybrid Single Particle Lagrangian Integrated Trajectory (HYSPLIT) model (fig. S5) originate from the south, especially the southeast . Low ffCO2 at WAO is observed during opposite conditions, such as during the summer months and when the wind direction is northerly (from the ocean). A ranking of the importance of each independent variable for the model training is shown in fig. S6. These results collectively indicate that the trained machine learning model provides a physically realistic representation of ffCO2 at WAO.It is, however, necessary to consider the performance of the machine learning algorithm and its associated uncertainty, which we assess as follows. First, we evaluate the performance of the model, as shown in 2, to the trained data from the original model for the period 1 January to 31 December 2019. Using the model to predict ffCO2 during previous nonpandemic years does not result in erroneous COVID-19–type signals, which should occur if the model prediction is consistently prone to overestimation.Second, we recognize that the model prediction is less robust than the trained model, and an additional uncertainty . Although the choice of start date does have a small impact on the magnitude of the pandemic ffCO2 signals, all predictions still show similar patterns associated with the two COVID-19 lockdown periods with a period of lockdown easing during summer 2020. None of the differences are outside of the uncertainties, except for during the first few months, when the uncertainties are small. The sensitivity of the random forest prediction to other parameters, such as the RAPO value used to calculate ffCO2 from APO, and the stiffness of the APO baseline fit were also tested, but no notable differences were found.Third, to ensure that the random forest prediction is not overly sensitive to the data at the beginning of the period , we examined ffCO2 data instead of APO data  in biospheric CO2 emissions. It is more likely that the initial reduction in CO2 during the late summer of 2020 is caused by an anomaly in biospheric CO2 fluxes, perhaps caused by the heatwave in the late summer; hence, no quantitative assessment of the impacts of COVID-19 on ffCO2 can be made using the atmospheric CO2 data.We also applied the random forest machine learning algorithm to WAO atmospheric COAPO data  but foun2 variations, to quantify the reduction in ffCO2 associated with the COVID-19 lockdowns in the United Kingdom during 2020–2021. Detection of COVID-19 signals is only possible owing to the high frequency of the APO measurements  and when combined with a machine learning algorithm that accounts for the influence of atmospheric transport variability on the observed ffCO2. It should be noted that although the machine learning algorithm allows the comparison of ffCO2 in the atmosphere between different years, it does not account for how atmospheric mixing translates CO2 fluxes into atmospheric mole fractions. We also acknowledge that training a statistical model for predictive purposes using trending data is likely to be inappropriate unless the trend is first removed. For our study, while there has been a decreasing trend in U.K. ffCO2 emissions since 2010, this trend is not visible in the APO-based ffCO2 at WAO  ffCO2 values. In this case, the timely detection of COVID-19 signals at WAO is only made possible by the availability of a continuous ffCO2 tracer combined with a method to remove the effects of atmospheric transport on the ffCO2 signal, such as the machine learning algorithm we have applied here.It might be feasible to replicate our machine learning–based analysis using a discrete, low-frequency ffCO2 from March to December 2020, is in broad agreement with three U.K. bottom-up emissions estimates, based on indirect energy and activity data. We refer to atmospheric measurement-based top-down estimates as direct and bottom-up estimates as indirect, because what matters from a climate change perspective is the change in radiative forcing in the atmosphere, caused by changes in atmospheric greenhouse gas concentrations, which top-down methods are able to measure directly. Our APO-based analysis is able to resolve the two U.K. lockdown periods individually, which are separated by a period of recovery during the summer of 2020 when lockdown restrictions were eased. This APO analysis, using data from a single U.K. measurement station, indicates that a network of continuous APO measurement sites would have strong potential for providing top-down estimates of ffCO2 emissions at regional scales, which corroborates the results of a recent modeling analysis  ratios in per meg units, rather than mole fractions, because O2 is not a trace gas, and its mole fraction is therefore affected by small changes in other gases, such as CO2 (WAO is situated on the north Norfolk coast (53°N) in a rural part of the United Kingdom. Atmospheric O2 and CO2  and fossil fuel combustion . Our calculation of APO is approximate because it does not account for the influences of CH4 and CO  in units of parts per million at WAO using hourly averaged APO data as shown in BL, the hourly APO “baseline” values , was determined using a statistical baseline fitting method  from the “COFFEE” (CO2 release and oxygen uptake from fossil fuel emission estimate) database  around the mean STILT-COFFEE RAPO value of 0.37 as a proxy for uncertainty in RAPO at WAO, giving an absolute value of ±0.11. The relative uncertainty in RAPO is therefore ±31% . The uncertainties of both αF and the O2:CO2 molar ratio of terrestrial biosphere-atmosphere exchange are included in our total RAPO uncertainty as per Uncertainty in the gridded fossil fuel emission ratio estimates of 2[APO] uncertainty to be calculated using the rules of error propagation. We estimate a mean hourly ffCO2[APO] uncertainty of ±42%, which is dominated by our estimate of uncertainty in RAPO, followed by our estimate of uncertainty in APOBL. At WAO, during the period 2010–2021, the mean ffCO2[APO] detected is 3.4 ± 1.4 ppm. We reiterate that our estimate of RAPO uncertainty, while being the largest term, is only a proxy for the actual uncertainty owing to a lack of available information and that future work on using APO to quantify ffCO2 should focus on refining RAPO values and their uncertainties. Uncertainty in APOBL can likely be reduced with denser networks of APO data, which may allow APOBL to be calculated from other sites, instead of having to rely on statistical fitting methods that likely have a higher uncertainty.Converting the absolute uncertainty of each term in 2[APO] dataset, we use the “rmweather” R package (version 0.1.51) (To account for the influence of atmospheric transport on our ffCOhttps://ready.noaa.gov/HYSPLIT.php) model (k-means clustering and the openair package in R (https://davidcarslaw.github.io/openair/). For the dependent variable, we use hourly ffCO2[APO], calculated using Using rmweather, we train a random forest model of 300 trees at WAO for the period 2010–2019 using 10 independent variables: hourly meteorological observations , which are measured in situ at WAO; temporal factors ; hourly atmospheric radon-222 activity, a tracer for atmospheric mixing that has been measured at WAO since April 2018 using an Australian Nuclear Science and Technology Organisation monitor  of the UEA estimate (We compare our COVID-19 signal detected from ffCO


To investigate the association between insufficient maternal gestational weight gain (GWG) during dietary treatment, and neonatal complications of small-for-gestational-age (SGA) infants born to mothers with Gestational diabetes mellitus (GDM).A retrospective case-control study was conducted, involving 1,651 infants born to mothers with GDM. The prevalence of a perinatal outcome and maternal GWG were compared among SGA, adequate- (AGA), and large-for-gestational-age (LGA); association with birth weight and GWG was identified using Pearson's correlation analysis; binary logistic regression was performed to determine the odds ratio (OR) associated with SGA.P < 0.01). GWG and pre-partum BMI were lower among the SGA mothers with GDM group  than AGA  or LGA  mothers with GDM group. Binary logistic regression showed that siblings who were SGA  and preeclampsia  were associated with SGA, but not GWG below guidelines (P > 0.05). The risk of SGA (25.7 vs. 19.1 vs. 14.2%) and FGR (15.3 vs. 10.9 vs. 7.8%) was higher in GWG below guidelines group than those in GWG above and within guidelines group, the risk of low Apgar score (6.4 vs. 3.0 vs. 2.8%) was higher in GWG above guidelines group than that in GWG below and within guidelines group (P < 0.05).In total, 343 SGA, 1025 AGA, and 283 LGA infants met inclusion criteria. The frequency of SGA infants who were siblings (41.7 vs. 4.3 vs. 1.9%) and composite of complications (19.2 vs. 12.0 vs. 11.7%) were higher in SGA infants than in those in AGA or LGA infants group (both Our findings demonstrated that GWG above and below guidelines, compared with GWG within guidelines, had a higher risk of adverse infant outcomes. Our findings also suggested that GWG below guidelines did not increase the risk for SGA, though SGA infants had more adverse outcomes among neonates born to mothers with GDM. Gestational diabetes mellitus (GDM), a common pregnancy complication, affects 1–28% of all pregnancies. Maternal GDM contributes to short- and long-term health risks both for mothers  and children  , 11.5–16 kg for normal women (BMI 18.5–24.9 kg/m2) and 7–11 kg for overweight women (BMI 25–29.9 kg/m2), 5–9 kg for obese women (BMI ≥30 kg/m2), and physical activity for at least 30 min/day were recommended.The inclusion criteria were the followings: (1) mothers were diagnosed with GDM with singleton pregnancy and live birth; (2) mothers were older than 18 years; (3) mothers/infant pairs with complete maternal  and neonatal data ; 4) all newborns were given routine examinations. The exclusion criteria: (1) infants presented with chromosomal aberrations, and genetic syndromes; (2) mothers with concomitant pathology that could affect glucose metabolism ; (3) mothers weren't given regular obstetric examinations. Only one birth infant for mother was included in our study. GDM was diagnosed according to guidelines for the diagnosis and treatment of GDM in China . The guiAll infants were divided into the following three groups: SGA infants born to mothers with GDM, AGA infants born to mothers with GDM, and LGA infants born to mothers with GDM. The diagnostic criteria of SGA infants: infants whose birth weight is below the 10th percentile at gestational age. The diagnostic criteria of AGA infants: infants whose birth weight is between the 10th and 90th percentile at gestational age. LGA was defined by a birth weight >90th percentile at gestational age. Classification of all newborns was defined according to the Fenton growth chart. GWG was defined as the difference between the final weight before delivery (within the last week before giving birth) and the pre-pregnancy weight (within 3 months before pregnancy). Based on the guidelines for maternal GWG of IOM, GWG was divided into the following three groups for analysis: GWG below guidelines (insufficient GWG), GWG within guidelines (sufficient GWG), and GWG above guidelines (excessive GWG). We performed matching according to neonatal gestational age (difference was ≤ 3 days), sex, and maternal age (difference was ≤ 3 years).∧9/L in the first 72 h of life. Fetal growth restriction (FGR) was determined by ultrasound, early FGR was defined as the gestational age was < 32 weeks and the following criteria were present: estimated fetal weight (EFW) or abdominal circumference (AC) below the 3rd percentile for the gestational age or absent end-diastolic flow in umbilical artery (UA); late FGR was defined as the gestational age was>32 weeks and the following criteria were present: EFW or AC below the 3rd percentile for the gestational age. Hypoglycemia was defined as blood glucose < 35 mg/dl or plasma glucose < 40 mg/dl within the first 48 h of life. Low Apgar score was defined as a 1-min Apgar score ≤ 7. Necrotizing enterocolitis (NEC) was diagnosed according to Bell criteria. Neonatal respiratory distress syndrome (NRDS) was diagnosed by (1) evidence of respiratory failure, (2) administration of exogenous pulmonary surfactant (3) radiographic evidence. Symptomatic polycythemia was defined as venous hematocrit >65 % or hemoglobin >220 g/L. Prematurity was defined as gestational age at birth of < 37 weeks. The presence of at least one neonatal complication  was assessed.Each mother/infant pair's demographic data, intervention condition, and medical information were collected individually using medical records. All mother/infant pairs underwent structured medical examinations and physical examinations. The following data on siblings of included infants also were recorded: maternal blood glucose after a 75-g OGTT, glycated-hemoglobin (Hb) level, gestational age at delivery, sex, birth weight and height of newborn, and mode of delivery. The weight and length/height of mother/infant pairs were measured by trained nurses using standard anthropometric methods, and pre-pregnancy weight was obtained according to maternal self-report. BMI was calculated by dividing weight in kilograms by the square of height in meters. Early thrombocytopenia was defined as a platelet count of < 150 × 10Data entry was performed by a trained clerk and a supervisor. Data were cross-checked by the co-author (VF) for any errors and discrepancies. Data entered incorrectly will be examined and corrected by the supervisor after confirmation with the participants or their obstetric records. Any revision of the original data will be tracked in detail.t-tests, ANOVA or kruskal-wallis test were used to analyze the differences in continuous variables. A two-sided chi-squared or Fisher's exact test was used for categorical variables presented as numbers and percentages. Binary logistic regression was used to determine risk factors for SGA infants born to mothers with GDM after adjusting for potential confounding variables. The association between birth weight and maternal GWG, pre-partum period BMI, and pre-pregnancy BMI was identified using Pearson's correlation analysis. Differences were considered statistically significant at a two-sided P value of < 0.05.Sample size calculation was performed using PASS 15. With 90% power and the assumption of relative risk = 2.0, we calculated that 256 SGA infants born to mothers with GDM were needed. Data were analyzed by using SPSS version 22 . Means (standard deviations) or median (range) was used to describe continuous variables; Overall, the final sample consisted of 1,651 mother/infant pairs. Of these, 343 SGA infants born to mothers with GDM, 1,025 AGA infants born to mothers with GDM, and 283 LGA infants born to mothers with GDM. There were 92 (5.6%) premature births, with 8 (2.3%) in the SGA infants born to mothers with GDM.2) than that in AGA  or LGA  mothers with GDM group (both P < 0.001). Preeclampsia was higher in SGA mothers with GDM than that in AGA mothers or LGA mothers . The incidence of primiparity (61.5%) and GWG below guidelines (50.4%) were highest in the SGA mothers group, and LGA mothers had the highest proportion of cesarean section (41.4%) and GWG above guidelines (39.2%).Relevant characteristics of mothers are presented in P < 0.01). The risk of symptomatic polycythemia was more frequent in LGA infants than that in the AGA or SGA infants group . The risk of early thrombocytopenia was increased in SGA infants compared to that in AGA infants (1.7 vs. 0.2%). However, there were no significant differences in the risk of Apgar score, hypoglycemia, hypoglycemia encephalopathy, NEC, NRDS, or perinatal death among the three groups .P < 0.01).Pearson's correlation analysis was used to determine the correlation between maternal factors and birth weight, and the results are shown in P < 0.05). There were no significant differences in adverse women outcomes, early thrombocytopenia, NEC, hypoglycemia, or NRDS among these three groups.Neonates were further grouped by the guidelines for maternal GWG of IOM as follows: GWG below guidelines, GWG within guidelines, and GWG above guidelines. Next, multivariate binary logistic regression analysis was performed to identify factors associated with SGA, and the results are shown in P > 0.05)  .We conducted a population-based, case-control study to explore the association between dietary intervention and the frequency of SGA. The main results of our study suggested the following:  siblingsDietary advice and exercise interventions are the first-line therapy for pregnant women with GDM. Dietary advice and exercise interventions alone for the prevention of GDM have been widely assessed; dietary and exercise interventions have been proven to reduce GWG in these studies , 16. So Dietary advice and exercise interventions can reduce GWG, Pearson's correlation analysis also showed that GWG was correlated with infants' birth weight in our study; but whether they can increase the incidence of SGA remains controversial , 15. IOMMoreover, some previous studies reported that excessive GWG (GWG above guidelines) had increased the odds of hypertensive disorders of pregnancy, cesarean delivery, macrosomia, and LGA , 36; butThe recurrence risk among siblings is widely used to measure shared genetic contributions. The frequency of SGA siblings in SGA infants born to mothers with GDM group was 10-fold higher than that in the AGA group in our study. Therefore, our findings demonstrated that genetic factors and preeclampsia were risk factors for SGA infants born to mothers with GDM. Recently, Spanish scholars indicated that smoking and neonate prematurity were also risk factors for these infants ; howeverOur study presents novel information, not previously reported, as we considered the frequency of siblings who were SGA, neonatal early thrombocytopenia, and abnormal GWG among GDM women. Our findings are also helpful for clinicians to design more accurate care programs for pregnant women with GDM. Our study also has several limitations. First, the number of SGA infants born to mothers with GDM may be small. Second, the other limitation is reliance on self-reported pre-pregnancy weight. Third, the time for the blood routine examination and blood glucose test was varied for each infant, which might affect the results. Fourth, it was a retrospective observational study; therefore, selection and information bias cannot be excluded.In conclusion, our findings suggested that GWG below guidelines did not increase the risk for SGA, though SGA infants had more adverse outcomes among neonates born to mothers with GDM. However, SGA infants born to mothers with GDM remain a major concern. Moreover, we identified that genetic factors and pre-eclampsia could be the risk factors for SGA infants born to mothers with GDM. Our research also demonstrated that GWG above and below guidelines, compared with GWG within guidelines, had a higher risk of adverse infant outcomes. These findings suggest that it is necessary to maintain a reasonable GWG among pregnant women with GDM to reduce adverse perinatal complications.The original contributions presented in the study are included in the article/supplementary material, further inquiries can be directed to the corresponding authors.The studies involving human participants were reviewed and approved by Ethical Committee in Guangzhou Women and Children's Medical Centre of Guangzhou Medical University, Number 2014121402, date of approval 12 December 2014. The patients/participants provided their written informed consent to participate in this study. Written informed consent was obtained from the individual(s), and minor(s)' legal guardian/next of kin, for the publication of any potentially identifiable images or data included in this article.DH, ML, BX, WZ, and JC designed the study, designed the data collection instruments, collected data, carried out the initial analyses, drafted the initial manuscript, and created the tables and figures. SC, YX, and JC conceptualized and designed the study, coordinated and supervised data collection, and helped draft the initial manuscript. HL, LW, PP, DY, YY, and JY designed the data collection instruments, collected data, and conducted the initial analyses. All authors contributed to the manuscript's critical revision, read, and approved the submitted version.


Birth weight is defined as the first weight of the newborn, ideally measured soon after birth. A recent Ethiopian survey estimated that 48% of births took place in health facilities. Data for women exposed to intimate partner violence (IPV) may be lacking in official statistics because these women may prefer to deliver at home, where data from non-institutional births, including reporting of birth weights, are not routinely recorded.The aim of this study was to investigate the association between maternal exposure to IPV during pregnancy and birth weight in a community in the Wondo Genet district of southern Ethiopia.We carried out a community-based prospective cohort study from February to December 2017. We followed up with 505 pregnant women and their newborns until after delivery. An interview about partner violence was done during pregnancy at home when enrolled. Field assistants who visited the homes measured the birth weight of each baby in grams. Twins and late birth weight measurements were excluded. Factors associated with birth weight were assessed by multiple linear regression.Birth weight was assessed within 48 h for 477 (94.5%) newborns and between 48 and 72 h for an additional 28 (5.5%). There were 365 (72.3%) institutional deliveries. In an adjusted regression analysis (IPV adjusted for socio-economic status), birth weight was 203 g lower (B −203 95% CI −320 to −87) among newborns of women exposed to IPV than among the unexposed. Birth weight was also lower in girls than in boys, in newborns delivered at home rather than in a health facility, and in babies with a younger gestational age.Maternal exposure to IPV during pregnancy was associated with lower baby birth weights. Antenatal clinics should consider routinely identifying IPV-exposed women, and identifying babies with lower birth weights at home is an important indicator. The birth weight of a fetus is an indicator of intrauterine growth and is the result of interactions between genetic, environmental, and social factors . Birth wDoes intimate partner violence (IPV) during pregnancy harm the baby? A systematic review and meta-analysis of studies conducted mainly in high-income countries showed that women who experienced IPV during pregnancy had an 18% higher chance of having a baby with LBW . A largeIn Ethiopia, the proportion of newborns with LBW has been reported to be between 8 and 28% , 15. A DAn earlier study in Ethiopia showed a 37% increased risk of LBW in pregnancies exposed to IPV; however, this study was only undertaken at two hospitals . Women eThe study was conducted in the Wondo Genet district located in the former Sidama zone of the Southern Nations, Nationalities, and Peoples Region (SNNPR). Based on the 2007 census and an annual population growth rate of 2.7% , the disThis was a community-based prospective cohort study investigating IPV. It was conducted between February and December 2017 among pregnant women who were enrolled at gestational age 25–34 weeks as listed by health extension workers  in two uThe sample size to investigate the association between IPV and birth weight was calculated using OpenEpi version 3.03 software . The totPregnant women living in the selected kebeles were enrolled through home visits. The sites were identified as being “urban” or “rural” according to the Ethiopian DHS definition of these terms . The selThe main outcome variable was birth weight measured in grams. The main exposure variable for this study was IPV during pregnancy, assessed in a home visit at enrollment using questions adapted from the WHO multi-country study questionnaire on women's health and domestic violence against women . In the Other covariates included the mother's age (years), educational status of the mother and her partner (no education/primary/secondary and above), monthly income (Ethiopian Birr), residence , prior history of preterm birth and stillbirth (no/yes), antenatal care at least one visit (no/yes), smoking (no/yes), and regular alcohol use (no/yes) by participants or their partner. The Edinburgh Postnatal Depression Scale (EPDS) was used to measure maternal depression , and hasThe data were collected between February and December 2017. The data used to achieve the objectives of the present study are the same as described for the previously mentioned survey of IPV in the study area . Other selected determinants were also studied. Preliminary analyses ensured that there was no violation of the assumption of normality, linearity, multicollinearity, and homoscedasticity. Complete case analysis was used for missing values to minimize potential bias. Variables having a correlation with birth weight at p < 0.2 levels and socio-economic and demographic variables were entered in the adjusted regression model. Maternal age, MUAC, and monthly income were entered as continuous variables. An adjusted regression coefficient (B) with a 95% confidence interval was reported. A p-value was considered statistically significant when < 0.05. Sensitivity analysis was also performed in a group of neonates whose birth weights were taken within the first 48 h after birth. We also analyzed risk factors for LBW  using logistic regression analysis (data not presented in a table).Data were analyzed using SPSS version 20  software. Chi-square and Fisher's exact tests were used to compare categorical variables. Mean values were compared using The study was conducted after obtaining approval from the Institutional Review Board (IRB) at the College of Medicine and Health Sciences, Hawassa University (Ref No: IRB/006/09) and regional ethical committee of Western Norway (Ref No: 2016/1908/REK vest). Permission from the parents as well as assent was obtained for those <18 years. In this setting, written consent was not culturally acceptable, but participants were comfortable with oral consent. Informed oral consent was obtained from each participant >18 years and recorded by the interviewer, according to the protocol approved by IRB. The study followed the ethical and safety guidelines recommended by the World Health Organization . All womOut of 589 pregnant women enrolled in the large follow-up study, we excluded 84 cases from the analysis: including eight women with twins, one preterm and two stillborns, 22 cases where the mother refused to have the baby weighed, 26 assessed after 72 h, two early neonatal deaths and 23 lost to follow-ups.Birth weight was assessed within 24 h for 234 (46.3%) newborns, between 24 and 48 h for 243 (48.1%) newborns, and between 48 and 72 h for 28 (5.5%) newborns. Mean birth weight by mother's characteristics is shown in Associations between birth weight and potential determinants including maternal exposure to IPV during pregnancy were analyzed by multivariable linear regression and are shown in Low birth weight  was found in 42 babies (8.3%). It was identified among 23 of the newborns of women exposed to IPV , compared to 19 among newborns of women not exposed to IPV . This represents a risk ratio of 4.3.Maternal undernutrition (thinness) and exposure to IPV during pregnancy were significantly associated with LBW in the multivariate logistic regression analysis after controlling for potential confounders. LBW was eight times more likely to occur in women exposed to IPV during pregnancy  and five times more likely in women with undernutrition (thinness) . Another factor associated with LBW was place of delivery, with around five times increased likelihood of LBW in home deliveries  than in facility deliveries, both crude and adjusted. The result did not change when analyzing the effects of interaction between IPV and nutrition (not shown in Tables).In this community-based follow-up study of 505 mother–singleton infant pairs in southern Ethiopia, the babies from IPV-exposed pregnancies were on average almost 200 g smaller. Among the women who experienced IPV during pregnancy, one in five gave birth to babies with low birth weight, whereas one in 20 unexposed women gave birth to babies with low birth weight.via related risk factors, such as the use of alcohol, resulting in impaired fetal growth , the study that was conducted in rural Ethiopia reported a LBW of 7.9% . But, itThis study had several strengths. A reasonably large study population of pregnant women was enrolled consecutively in the area, and the participants were, as far as we know, representative for the area. This ensured a good representation of rural women and home births.The study also had some limitations. A few pregnant women may have been missed in the households, including women who try to hide their pregnancy, and we have no information about this. The birth weight was missing for 14% of the women including 4% who were lost to follow-up. However, there were no large differences in their background characteristics, and this likely represented no major selection bias. The proportion of LBW was slightly higher (8.3%) among those measured at 0–72 h than when measurements were limited to 0–48 h (7.7%), but this had little impact on the association between IPV and birth weight in our sensitivity analysis see . An earlWomen exposed to intimate partner violence during pregnancy tended to have smaller babies than unexposed women. Birth weight was also associated with the sex of the newborn, gestational age at birth, and place of delivery. Antenatal clinics may consider routinely identifying women suffering from IPV and then referring them to the relevant organizations. The positive efforts by the Ministry to enhance facility delivery should include identifying babies with lower birth weights who need special care.The original contributions presented in the study are included in the article/supplementary files, further inquiries can be directed to the corresponding author/s.The studies involving human participants were reviewed and approved by Institutional Review Board (IRB) at the College of Medicine and Health Sciences, Hawassa University (Ref No: IRB/006/09) and regional Ethical Committee of Western Norway (Ref No: 2016/1908/REK vest). Written informed consent to participate in this study was provided by the participants' legal guardian/next of kin. Informed oral consent was obtained from each participant > 18 years and recorded by the interviewer.SB conceptualized, designed and implemented the study, analyzed the data, and drafted the manuscript. AA and SGH contributed with planning, analyzed the data, revised and edited the manuscript, and supervised. All authors have read and approved the final manuscript.This research was funded by the NORHED/SENUPH , Agreement no. ETH13/0025. The funder had no role in study design, data collection and analysis, decision to publish, or preparation of the 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.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


This study investigates the role of maternal decorin levels measured in the early second trimester of pregnancy in detecting the potential for preterm birth in late pregnancy.The prospective, case-control study was carried out in tertiary university hospital from June to December 2021.Maternal serum samples were collected from 350 women aged 18–40 years with a singleton pregnancy during early second-trimester screening for aneuploidy, and stored at -80°C. All participants were followed up until delivery and 25 patients diagnosed with preterm birth group and 40 full-term healthy women were included in the study.The median maternal serum decorin level was 3.82 (1.15–12.37) ng/ml in the preterm birth (PTB) group and 4.63 (1.20–10.02) ng/ml in the control group, there was no statistically difference between the groups (p = 0.111). The mean gestational age was statistically significantly lower in the preterm birth group (33.1±2.7 weeks) than in the control (39±1.16 weeks) (p <0.001). The mean fetal weight was statistically significantly lower in the preterm birth group (2023.8±477 g) than in the control group (3309.7±353 g) (p <0.001).Early second-trimester serum levels of decorin alone may not be a sufficiently accurate biomarker as a biochemical model for the prediction of preterm birth in asymptomatic women. Preterm birth is defined as birth before 37 weeks of gestation. As one of the leading causes of perinatal and neonatal morbidity and mortality in many countries around the world, it is considered a major obstetric and health problem. The prevalence of preterm birth is higher than 11% in many regions, and ranges from 5–7% in northern European countries, indicating clear differences in regional preterm birth rates.2While various spontaneous or iatrogenic mechanisms can lead to preterm birth, 75% of cases are spontaneous.4The pathogenesis of preterm birth is still not fully understood, although the results of previous studies indicate a complex syndrome with multiple causes. PPROM, short cervical length at 23–24 weeks of gestation, positivity for fibronectin, and history of PPROM and connective tissue disorders associated with connective tissue damage, such as Ehlers Danlos Syndrome (EDS) that can cause a weakening of fetal membranes, may serve as guides for the early detection of preterm birth.), a member of the small leucine-rich proteoglycans (SLRPs), is found in the extracellular matrix (ECM). As a molecule that is produced mainly by reproductive tissues, Decorin is, together with biglycan, the main proteoglycan forming the intermediate and reticular layers of human fetal membranes.6Decorin  reported maternal biglycan levels to be high in early pregnancy, while decorin levels were low in the group of pregnant patients who went on to develop PPROM in late pregnancy.7The present study investigates the clinical significance of decorin levels measured in the early second trimester of pregnancy, exploring its potential in the determination of preterm birth developing in late pregnancy.This single-center, prospective, case-control study was carried out a tertiary university hospital – between June and December 2021. Written informed consent was obtained from each participant prior to their involvement in the study. The study protocol was approved by the hospital’s local Ethics Committee (Date: 2011-KAEK-25 2021/08-03).In the study, maternal serum samples were collected from 350 women aged 18–40 years with a singleton pregnancy during second-trimester screening for aneuploidy between (15+0–19+6) weeks of gestation. The patients’ medical history, age, height, weight, gravidity and parity, and history of gestational hypertension (GHT), preeclampsia (PE), fetal growth restriction (FGR), PPROM and preterm delivery as adverse pregnancy outcomes were recorded. The pregnancy follow-ups, peripartum outcomes and delivery details of all participants were retrieved from the hospital database systems and the records of the delivery clinics. Our study included a total of 65 patients, including 25 pregnant women diagnosed with preterm birth based on pregnancy and delivery details, and 40 full-term healthy pregnant women with no complications during pregnancy follow-up, delivery or the postpartum period. The patient and control groups were matched for age and BMI.Preterm birth is defined by the World Health Organization (WHO) as a live birth occurring before 37 weeks of gestation, while PPROM refers to the premature rupture of fetal membranes before the onset of labor.8Pregnant women with multiple pregnancies, incompetent cervix, congenital anomalies, intrauterine growth retardation (IUGR), intrauterine fetal demise or gestational hypertensive diseases, and those using acetylsalicylic acid or anticoagulants were excluded from the study. Other exclusion criteria were pre- and gestational diabetes mellitus, polyhydramnios, connective tissue diseases, vasculitis, chronic kidney and/or liver disease, hyper/hypothyroidism, hematological disorders and smoking, and a previous pregnancy history of chorioamnionitis, PPROM and preterm birth.2), gravidity and parity were recorded for each patient at the time of sampling. Gestational age was calculated according to the first day of the last menstrual period, and was confirmed by measuring the first-trimester crown-rump length on obstetric ultrasonography.Demographic data such as maternal age, BMI .The venous blood samples collected from all study patients at the early second-trimester screening for aneuploidy (15–20 weeks) were centrifuged at 3500 rpm for 10 minutes and then stored at -80Data were analyzed by SPSS 22.0 for Windows  statistics program. Shapiro-wilk test was used to determine the normality of distribution of variables. Descriptive data were expressed as mean ± standard deviation for normally distributed variables while they were expressed as median (minumum-maximum) or non-normally distributed ones. Categorical variables were shown as percentages and frequency. Student t-test was used for variables showing normal distribution between two groups and Mann-Whitney U-test was performed to compared non-normally distributed variables between two groups. Categorical variables were compared between groups by Pearson Chi-square and Fisher’s exact test. Alfa value <0.05 was considered as statistically significant.The study, conducted with 350 pregnant women, included 25 patients diagnosed with preterm birth (7.15%) in which preterm delivery occurred following PPROM in 17 cases (68%) and preterm labor in eight cases (32%).The mean age of the patients was 28.2± 6.6 and 27.7±4.1 years in the preterm birth and control groups, respectively, with no statistical difference between the groups (p = 0.756). The mean body mass index was 28.2±6.1 in the PTB group and 29.2±4.5 kg/m2 in the control group (p = 0.11). There was no statistically significant difference between the two groups. Similarly, there was no difference between the groups median gravidity and parity numbers (respectively p=0.154 and p=0.853). The median gestational age at the time of serum sampling was 16.35 weeks for the preterm birth cases and 16.85 weeks for the controls (p = 0.177). The demographic data of the study participants are presented in The mean maternal serum level of decorin was 3.82 (1.15–12.37) ng/ml in the PTB group and 4.63 (1.20–10.02) ng/ml in the control group, with no statistically significant difference between the two groups (p = 0.111). There was no significant difference in the decorin levels of the preterm labor and PPROM groups (p=0.344). The mean gestational age at birth was statistically significantly lower in the PTB group (33.1±2.7 weeks) than in the control group (9±1.16 weeks) (p <0.001). The average fetal weight was statistically significantly lower in the PTB group (2023.8±477 g) than in the control group (3309.7±353 g) (p <0.001). Similarly, the one minute and five minutes Apgar scores were lower in the PTB group than in the control group (p <0.001 for both parameters). Clinical characteristics of the study groups are summarized While 64% had a Caesarean delivery and 36% had a vaginal delivery in the preterm birth group, the rate of acute fetal distress was 48%. The number of newborns followed up in the neonatal intensive care unit was 13 (52%) .Preterm birth places a great financial burden on society due to the resulting prematurity, and the need for neonatal intensive care and specialized health personnel. The development of supportive therapies and facilities also increases neonatal morbidity, thereby leading to the prolonged life expectancy of newborns and potential comorbidities.Preterm birth can occur for various reasons, such as inflammation, hemorrhage, activation of the maternal or fetal hypothalamic-pituitary axis, immune dysregulation, distension of the myometrium and cervical insufficiency.Decidualization involves extensive changes in the ECM, indicating the transformation of fibroblasts into secretory cells in the endometrial stromal tissue. In pregnancy, this process is associated with the remodeling of the spiral arteries, as well as the transformation of the uterine glands and uterine natural killer (uNK) cell flow, ensuring adequate blood supply to the growing fetus. Decorin is found in the cervix, uterus, fetal mesenchymal cells, and placental decidua in humans at varying levels, depending on gestational age. Decorin binds to many components of the ECM, growth factors and receptor tyrosine kinases, and is required, together with N core proteins, for collagen fibrillogenesis and for increasing fibril diameter and tensile strength.13Decorin also influences the movement of cells by modulating the interaction between membrane surface receptors and matrix protein ligands. While decorin-expressing cells have less migration ability, they also accumulate matrix fibronectin. There have been numerous studies reporting that the amount of both decorin and collagen decreases as the membrane tensile strength reduces before birth in human fetal membranes.In their model study of mice, Calmus et al. reported biglycan and decorin to be necessary for the maintenance of an uneventful pregnancy until term, and to have a gene dose-dependent and compensatory effect on each other.15The 2014 study by Meinert et al. revealed the role of biglycan and decorin in the maintenance of pregnancy in an inflammation setting, and also in the fetal membrane response to inflammation.16Various studies have shown that decorin concentrations decrease in fetal membranes immediately after a term birth, while the levels decrease in amniotic fluid in preterm and full-term deliveries. This decrease in decorin during the third trimester and delivery may contribute to the mechanical weakening, leading to the onset of labor.17Horgan et al. reported a significant degradation of decorin in the amnion in chorioamnionitis, and established a decreased fetal membrane decorin expression in preterm birth with PPROM when compared to preterm birth without PPROM only in the presence of infection. The authors suggested that the low-level regulation of decorin due to inflammation might be responsible for the weakened fetal membranes in PPROM.18Atalay et al. found serum decorin levels to be significantly lower in their preterm labor group than in their healthy control group in their study of women with 24–32 weeks of pregnancy presenting with preterm labor. The authors reported that while serum decorin levels alone could predict preterm birth before 37 weeks of gestation, a combined measurement with cervical length was required to predict preterm birth within 7 days and before 34 weeks of gestation.19Underhill et al.’s first study in 2019 reported maternal biglycan levels to be high in the early second trimester, and decorin levels to be low in the group of pregnant patients who went on to develop PPROM in late pregnancy.7A mice model study by Pantham et al. found that the loss of ECM proteins biglycan and decorin led to marked changes in the transcriptomic profile of fetal membranes during gestation, as well as possibly cell cycle suppression and activation of the complement system. This resulted in a lack of proliferation in fetal membrane mesenchymal cells, and ultimately a loss of fetal membrane integrity and premature birth.20In the light of the data provided in the various studies mentioned above, our study followed pregnant women whose maternal serum samples were collected in the early second trimester, and compared the maternal serum decorin levels between selected healthy patients delivered at term and those with a preterm delivery to determine the clinical significance of decorin in the prediction of preterm birth.We identified a lower DCN level in the preterm birth group (3.75 ng/dl) than in the healthy control group (4.60 ng/dl), although the difference was not statistically significant (p = 0.111). When the preterm births were classified into two subgroups as PPROM and Preterm labor, based on the etiology, no significant difference was identified between the serum decorin levels of the groups (p = 0.344). Similar to the present study, Mennella et al. reported that decorin levels in healthy pregnancies remained stable with a decrease of less than 1% per week throughout the pregnancy, and that serum decorin and biglycan levels did not exhibit abnormal expression patterns in early pregnancy before the onset of preterm labor or PPROM. These findings support the lack of difference in decorin levels between the preterm birth and healthy control groups reported in the present study.21It is single-center design and the relatively small sample size. The study was further limited by the fact that we were unable to evaluate decorin levels in the weeks of delivery in the PTB and control groups, and that no investigation was made into the presence of infection. Furthermore, cervical length was not measured in the study group patients in the early second trimester when the maternal serum samples were collected, which is another limitation of the study.Molecular new markers play an important role in research and clinical practice. However, for the prediction of preterm birth in asymptomatic women based on the results of the present study, early second-trimester serum levels of decorin alone may not be considered a valuable biomarker as a biochemical model. That sad, the results still have the potential to open new doors on a new approach to the identification of women at high risk of preterm labor and PPROM, although large, prospective studies will be required to reach definitive conclusions.


This US-based, prospective observational cohort study evaluated the safety of a quadrivalent inactivated influenza vaccine  in pregnant persons immunized over four influenza seasons between 2017 and 2021. Pregnancy outcomes included live birth, stillbirth, spontaneous abortion, and elective termination. Infant events of interest were major congenital malformations (MCMs), preterm birth , and low birth weight (LBW). Data were descriptive; prevalence point estimates were reported with 95% confidence intervals (CI). A total of 483 pregnant persons were given IIV4 and evaluated; 477 (98.8%) reported a live birth, and there were 2 stillbirths, 4 spontaneous abortions, and no elective terminations or maternal deaths. The prevalence rates of infant events were as follows: preterm birth, 7.2% ; LBW, 5.4% ; and MCMs, 0.8% . Point estimates and upper 95% CIs of the observed prevalence rates were lower than or similar to background prevalence in the general US population. Our findings suggest no evidence of a safety concern with vaccinating this group at high risk of influenza complications and are consistent with published data from databases and surveillance systems that monitor the safety of influenza vaccines in pregnant persons. Influenza vaccination during pregnancy decreases the risk of influenza disease and its complications among pregnant persons and their infants . PregnanRecent estimates of influenza vaccination during pregnancy have shown that only 61% of pregnant people reported receiving an influenza vaccine, which is well short of the 80% Healthy People 2020 coverage target, and there is a coverage gap of ~15% between more and less advantaged socioeconomic groups ,10. A keA quadrivalent inactivated influenza vaccine  was first licensed for use in adults in 2016 in Australia and the US and has since been approved in New Zealand, Canada, Argentina, Germany, Austria, and South Korea. In this prospective observational study, we evaluated the safety of IIV4 in terms of adverse pregnancy outcomes and major congenital malformations (MCMs), preterm birth, and LBW in infants born to people who received IIV4 during pregnancy.This prospective observational cohort study, a postmarketing commitment to the Food and Drug Administration (FDA), was conducted over four consecutive influenza seasons from 2017–2021. The Syneos Health Registry Coordinating Center  served as the single site for the study, but data were collected from obstetrics/gynecology clinics that used IIV4 as a component of routine care and were willing to enroll pregnant people vaccinated with IIV4 in their clinic. Study participants could also self-enroll .The schedule of visits, laboratory tests, all treatment regimens, and timing of immunizations in this strictly observational study were determined by the treating health care provider (HCP). This safety study did not include clinical assessments to evaluate influenza vaccine effectiveness. The study was designed and conducted in a manner consistent with the 2002 US FDA guidelines for pregnancy exposure registries, which was effective when the study started , as wellThe study population included pregnant persons residing within the US who received IIV4 at any time as part of routine pregnancy care for whom sufficient data were available to confirm the subject was enrolled prospectively, the date and specified brand of IIV4 vaccination, pregnancy outcome, and whether an event of interest in the fetus or infant had occurred. Both passive and active recruitment strategies were implemented. Subjects were recruited by healthcare providers (HCP) from obstetrics/gynecology clinics that used IIV4 per routine care and were willing to enroll eligible and consenting pregnant persons vaccinated in their clinic. In addition, a targeted awareness campaign was undertaken that allowed subjects to self-enroll in the study. Subject selection and recruitment strategies are described in greater detail in the IIV4  is a purified, split, inactivated, egg-derived, nonadjuvanted, quadrivalent influenza virus vaccine formulated to contain hemagglutinin antigens from the A(H1N1), A(H3N2), B/Yamagata, and B/Victoria strains selected for each season by the World Health Organization (WHO). Vaccination was administered solely as part of routine clinical care at the discretion of the pregnant person and the treating HCP.Pregnancy outcomes of interest included live birth, stillbirth , spontaneous abortion , or elective termination. Ectopic pregnancy or molar pregnancy were reported as pregnancy types. Infant-related events of interest included preterm birth , low birthweight (<2500 g), or MCM, which was defined and coded using criteria specified by the CDC Metropolitan Atlanta Congenital Defects Program (MACDP). Specifically, any major structural or chromosomal defect or combination of three or more minor defects in liveborn infants, stillbirths, or fetal losses of any gestational age (including outcomes prior to 20 weeks’ gestation or birthweight <500 g) was considered an MCM. Information on clusters of minor abnormalities (as defined by the CDC MACDP) and data from aborted fetuses of less than 20 weeks’ gestation, when available, was included to increase the sensitivity of monitoring. Infant size was also calculated using the 2006 WHO Child Growth Standards used to derive the z scores for weight-for-age and sex with outputs reported in percentile categories .HCPs from participating clinics and/or pregnant persons or their legally authorized representatives served as data reporters for this study. The only data collected were those documented in the patient’s medical records as part of routine care; no assessments were required as part of the study. Data regarding events of interest were provided by the obstetric HCP and/or pediatrician attending the birth. Gestational age at birth and birthweight were used to assess preterm birth and LBW. A teratologist who was blinded to the timing of vaccination reviewed all reported congenital anomalies and classified them using the CDC MACDP system and provided an opinion regarding the gestational time window during which exposure to IIV4 could potentially have had an impact on the reported birth defect. A Scientific Advisory Committee (SAC), consisting of experts in the fields of obstetrics/maternal-fetal medicine, pediatrics, clinical research, infectious disease, epidemiology, and teratology, met periodically to review the data and make recommendations on the coding and classification of MCMs and other outcomes of interest.This was a descriptive study, and a planned sample size of 550  was not based on a statistical hypothesis to be tested but instead was determined based on the US background prevalence of MCMs (2.8%), preterm births (11.4%), and LBW (8.0%) according to the CDC MACDP and National Vital Statistics System (NVSS), as available at the time the study was designed ,20. AssuPregnancy outcomes and events of interest were summarized overall, by trimester of exposure, and by maternal age at conception categories; 2-sided 95% confidence intervals (CIs) were calculated for pregnancy outcomes using the exact binomial distribution (Clopper–Pearson method). The prevalence of MCMs was calculated by including all MCMs in liveborn infants as well as MCMs in stillborn infants with a gestational age of ≥20 weeks in the numerator and including all live births in the denominator. Overall and stratum-specific prevalence point estimates and 1-sided 95% CIs were calculated using the exact binomial distribution for prevalence of MCMs among all evaluable live births, for prevalence of LBW among evaluable pregnant persons with a singleton live birth without MCMs, and for prevalence of preterm birth among evaluable persons enrolled prior to 37 weeks’ gestation with a singleton live birth without MCMs. Birth outcomes were compared to the most recently reported corresponding prevalence estimates from the CDC MACDP and NVSS ,22.n = 6) or enrolled at one of five participating clinics in New York and North Carolina (n = 484)  . Influenn = 484) . The prin = 484) .The mean maternal age at conception was 29.8 ± 5.4 years . The stuOverall, 477 (98.8%) pregnancies ended in a live birth, including 97.1, 99.5, and 100% of pregnancies with exposure to IIV4 during the first, second, and third trimesters, respectively . Two stiThe infant primary analysis population for the calculation of the prevalence of MCMs included 485 liveborn infants from 483 pregnant persons. Nine carried twin pregnancies, of which one resulted in a spontaneous abortion, and there were two stillbirths and three spontaneous abortions among the singleton pregnancies. After excluding the 16 liveborn infants from the remaining twin pregnancies and four infants with MCMs, the LBW analysis population included 465 infants . The preThe infant outcomes appear in Our findings are consistent with previous studies looking at all influenza vaccines combined , showing that maternal immunization does not increase the risk of adverse pregnancy outcomes ,14,15,23In 2020, the NVSS reported a preterm birth rate of 10.1% in the US . The estAlthough the pregnancy registry was not designed or powered to investigate specific malformations, there were two cases of gastroschisis among the four reported MCMs that merit further discussion. Gastroschisis is the most common type of congenital abdominal wall defect in infants. Its prevalence has increased over the last three decades, with the highest prevalence observed among mothers <20 years of age  ,27,28,29Most registry participants were enrolled at one of five obstetrics/gynecology clinics in the eastern US, and a larger sample size would be required for full representation of the population of pregnant persons in the US. Despite these limitations, the study population included the most common racial and ethnic groups in the US, in similar proportions to the general population. In addition, the study population encompassed a broad range of maternal ages (18–43 years) and a relatively even distribution in the number (0 to ≥4) of previous pregnancies. Approximately 35% of registry participants were exposed to IIV4 in the first trimester, which is important for the assessment of maternal exposures and birth defects. High-risk pregnancies were not specifically excluded from the study, although there is a chance that very high-risk pregnancies may not have been under the prenatal care of the routine obstetrics/gynecology clinics from which the pregnant persons were enrolled. Particularly, in comparison to other pregnancy registries reported for branded influenza vaccines ,35, the A general limitation of noninterventional studies in which HCPs collect data during routine care is the potential for missing data, including a limited level of detail, lower precision of measures reported, and potential ascertainment bias of adverse pregnancy outcomes . Finally, this study did not use an internal comparator group but compared the findings in the exposed study population to historical population statistics from the general US population. Potential differences in the characteristics of the underlying populations cannot be ruled out. Although the study used the same case definitions for MCMs as the MACDP, there were important differences in data collection. As an active surveillance program, the MACDP includes all defects reported in the defined region, whereas participation in this study was voluntary. Furthermore, the MACDP counts any MCMs that are detected in an infant up to the age of 6 years, whereas follow-up in this study ended upon completion of the pregnancy. As a consequence, major MCMs not immediately detectable at birth might have been missed.The findings of this prospective safety study suggest that IIV4 exposure during pregnancy is not associated with any safety concerns based on comparison of the study results to background prevalence rates reported by the NVSS and MACDP. Our findings are consistent with published data from various databases and surveillance systems that monitor the safety of influenza vaccines in pregnant persons, further supporting use of influenza vaccines in this group at high risk from influenza.


BACKGROUND: Numerous birth weight standards for twins have been reported in western countries, whereas little is in Japan. The aim of this study is to present birth weight, birth length, chest circumference, and head circumference references, clarifying features related to these body size parameters, and to compare our birth weight references with recent report of birth weight norms of Japanese twins using the vital statistics.METHODS: The subjects consisted of 1,061 twin pairs in total, with birth years ranging from 1968 through 1990. Data was obtained from the Twins Protocol Questionnaire, which asked for information about twins’ growth and development in infancy, and the “Maternal and Child Health Handbook,” which was presented by Ministry of Health and Welfare. Statistical means, standard deviations, and selected percentiles by gestational age were calculated and smoothed using data that contained at least gestational age and one of the four items.RESULTS: Birth weight was significantly lighter than that of singletons when three additional parameters, especially chest and head circumference, were not measured. Gestational age was correlated with weight, length, chest circumference, and head circumference, in that order, for both sexes. Compared with singletons, birth weight difference in twins was marked and slight difference was observed as to length, whereas no difference was observed as to chest and head circumference. The present results as to birth weight were consistently similar to the birth weight norms of twins using vital statistics in Japan.CONCLUSION: Growth standards for twins, especially as to birth weight, are essential to understand and evaluate intrauterine growth of twins. Because birth weight is the strongest indicator of the risk of perinatal death, birth weight norms are important both for clinical practices and for epidemiologic studies.3 It is well known that intrauterine growth of twins differs from that of singletons. Many studies concerning birth weight standards by gestational age have been reported.3-14 Other body size parameters at birth, such as birth length,6,9 chest circumference,6,9 and head circumference6,9,12 have not been reported as consistently as birth weight, partly because these parameters are not routinely used as indicators of perinatal risk. Comparing these characteristics in twins with those of singletons may provide clues about the patterns of twin growth.Twins in Japan are known to be several times more prone to die perinatally than singletons.14 these norms are probably the most reliable ones in Japan. The limitations of these norms are that the data is calculated by 100g intervals, and the gestational ages are recorded in weeks, not days. Moreover, no norms are presented regarding birth length, chest circumference and head circumference at birth. The aim of this study is to compare our birth weight data with these norms and to present birth references for other body size parameters using a large-scale sample of normally developed Japanese twins.Previous studies have been conducted mainly in western countries. In Japan, relevant factors, such as the perinatal medical system, twins’ birth rates, and body size of mothers, differ significantly from western countries. Recently, twin birth weight norms in Japan based on birth records in the vital statistics were presented;The subjects consisted of 1,061 twin pairs , all of whom were applicants to the secondary education school attached to the faculty of education of the University of Tokyo from 1981 through 2002 (birth year ranged: 1968-1990). This school was established in 1948 and adopted a unique entrance system. Of the approximately 50 pairs of twins, 12 years of age and living in the Tokyo metropolitan area, who take an examination every year, about 15 pairs are admitted. All the parents of applicants must hand in a Twins Protocol Questionnaire, which gathers information on family structure, obstetrical findings of mothers, physical growth, zygosity, and motor and mental development of twins from birth through 11 years of age. One of the parents of each applicant, usually the mother, participates in a medical interview by two or three interviewers , in which their responses to the questionnaire are checked carefully. The data of this study are based on the records from the Twins Protocol Questionnaire and the “Maternal and Child Health Handbook,” presented by the Ministry of Health and Welfare for all pregnant women, including obstetrical records written by obstetricians. The data obtained by these procedures were accepted without modification; therefore, the method and accuracy of measurements could not be ascertained.15,16 and DNA/blood testing. All same-sex twins and their mothers answered the zygosity questionnaire, on the basis of which twins’ zygosity was determined with an accuracy of more than 95%. Zygosity diagnosis using genetic markers or DNA polymorphisms was performed for those twin pairs who were admitted to the school. The characteristics of the subjects are presented in Zygosity was determined by means of the questionnaireBirth complications of mothers or twins, for example placenta previa, placental abruption, coiling of the umbilical cord, neonatal asphyxia, growth discordant twins, and twin-to-twin transfusion syndrome, were observed to varying degrees. None of these were grounds for exclusion from the study. In general, it was very difficult to set clear and consistent inclusion/exclusion criteria, as more than ten percent of the present subjects had at least one of the complications mentioned above. Moreover, no subjects showed apparent growth retardation at 11 or 12 years of age. Surely this represents one of the largest and most thorough sets of accurate growth data on twins in Japan, especially because zygosity testing is very rare in Japan. Informed consent concerning statistical analysis of the subjects’ data was performed by written documents as part of the application process.First, factors that affect body size parameters at birth were confirmed by stepwise regression analysis, with a threshold significance level of 0.10. The variables considered were sex, birth order in twins (first-twin or second-twin), gestational age, maternal and paternal age at twin birth, intervals between marriage and twin birth, birth year of twins, parity, zygosity, and presentation. For qualitative variables, the following codes were used. Sex, female: 0, male:1; Birth order, first-born:1, second-born:0; Zygosity, monozygotic:0, dizygotic:1; Presentation, non-vertex:0, vertex:1.Next, body size parameters were analyzed according to the missing data conditions; for example, birth weight was analyzed whether or not birth length was measured.The effect of parity was analyzed according to sex. Next, the correlations of gestational age and body size parameters were analyzed.Reference body size parameters at birth were analyzed according to sex, considering the results of above mentioned procedures. Means, standard deviations, and selected percentiles by gestational age were calculated. In the process of calculating percentile values, the critical effect of outliers based on birth complications, if they existed, were expected to be excluded.17 and twins14 in Japan.Smoothing of growth curves was performed by a cubic spline function and compared with previously reported norms for the general population® for Windows.18 Smoothing of growth curves was performed using PROC TRANSREG program, by specifying the ‘pspline’ model.Statistical analyses were performed using SAS17 In this study, standards for birth weight, length, chest circumference, and head circumference by gestational ages from 22 to 41 weeks were determined by the analyses of birth size data on 1,133 infants whose gestational ages were confirmed by ultrasonographic techniques in early gestation with the cooperation of 21 major medical centers throughout Japan. The 10th, 50th, and 90th percentiles for each body size parameter according to sex are presented. For birth weight, parity was also considered.The general population standards  that we adopted were based on hospital data presented by a research group of the Ministry of Health and Welfare in 1998.14 These standards were calculated based on birth records in the vital statistics from 1988 through 1992, which included about 64,000 live birth twin individuals. The 10th, 50th, and 90th percentiles of birth weight by gestational ages from 24 to 41 weeks according to sex and parity are presented. No reliable standards for length, chest circumference, and head circumference were available for twins in Japan.The twins’ standards we adopted were the recently reported birth weight standards in Japan.2=0.31, p=0.0001 for birth weight, R2=0.30, p=0.0001 for length, R2=0.25, p=0.0001 for chest circumference, and R2=0.18, p=0.0001 for head circumference). Sex and parity also contributed to body size parameters, albeit slightly. The effect of parity was second largest (p=0.0001) influence on birth weight, though the effect itself was not so large (R2=0.02).The results of stepwise regression analysis showed that the contribution of gestational age was the strongest  and 134 (6.3%) respectively, whereas the number of missing values for chest circumference and head circumference were 283 (13.3%) and 281 (13.2%), respectively. Two possible reasons were (1) the avoidance of troublesome work, or (2) an emergency circumstance, in which there was no time to measure parameters other than weight and/or length. To confirm the latter possibility, birth weight or birth length was compared according to the missing data condition; the results are shown in Using subjects that had data on gestational age and all four parameters , we calculated correlation coefficients of body size parameters. Gestational age was correlated with weight , length , chest circumference , and head circumference , in this order for both sexes; the results were in accordance with the regression analyses. Birth weight showed high correlations with length  and chest circumference , and a slightly lower correlation with head circumference .17 and twins,14 considering the effect of parity. As the sample size below 32 weeks was small, smoothing was performed only for 33-41 weeks gestation. Smoothed 50th percentile curves are compared with singleton norms , 10.6% (30/283), and 10.7% (30/281), respectively. In cases in length, chest circumference, and head circumference were measured, the frequency of neonatal asphyxia was 5.1% (102/1988), 5.0% (91/1839), and 4.9% (91/1841), respectively. These findings strongly suggested that the reason for missing data was an emergency situation; these subjects may include a disproportionate number of unhealthy or at-risk infants. Conversely, chest or head circumference may only be measured for relatively healthy neonates. For these two parameters, it seems important to bear the possibility of positive selection bias in mind.As shown in 6,9 as to this parameter; moreover, it has been pointed out12 that length is unreliable, with many different measurement techniques. Nevertheless, the results of this study were in accordance with previous reports. The length of twins became larger with gestational age, and males were, on average, larger.As to length, a slight deficit in twins was observed. About a 1cm deficit was observed at 41 weeks for both sexes. Little has been reported12 who analyzed over 5,300 head circumferences of twins, differences in head circumferences between singletons and twins are only evident with gestations longer than 35 weeks, and from 37 weeks’ gestation onwards, the mean head circumference of singletons exceeded that of twins by only 5 mm. The deficit of head circumference in twins, though it exists, is certainly not so large as birth weight.Chest and head circumference were nearly the same as or even larger than those of singletons, partly suggesting the positive selection bias mentioned above. According to Bucker and Green,3 Besides, the difference between the results of most of the studies may partly result from the differences in study populations. Taking all these things in consideration, it seems very difficult to clarify the growth difference between singletons and twins meaningfully using the present sample.It is said that previous population-based and large hospital-based investigations do not agree with regard to the gestational age at which singleton and twin growth curves start to differ.23-25 Each method has its merits and demerits.It is important to recognize that often-reported body size standards by gestational age are cross-sectional and essentially different from real intrauterine growth or longitudinal measurements by ultrasound examination.10 because secular trends have been observed. In this study, secular trends of body size were not considered, partly because no strong birth year effect was observed by regression analysis.Birth years of the present sample were distributed roughly over twenty years. It has been recommended that birth weight norms should be updated every 5-10 years,In interpreting the results of the present study, the following selection biases should be considered. First, both twins are alive and have shown no marked growth disturbance through age 11 or 12. Second, the subjects are only in the Tokyo Metropolitan aria. Third, the subjects are all applicants to an entrance examination for a university-affiliated school. The direct effect of these selection biases is difficult to specify. The selection bias should be much lower in the population-based study using the vital statistics, which theoretically contain the data of most live-born twins of the year. We think the present data may offer reference birth size parameters for normal development through infancy, or the upper limit of intrauterine growth norms.Though there were several limitations in the present results, some of the features related to body size parameters at birth in twins were made clear. As shown in


Complex anal fistula with synovial cyst at the ischial tuberosity is a very rare condition. The key point in the operation is to protect the anal sphincter and completely peel off the cyst. Here, we report a successful case of closing the internal opening with the advancement flap technique and completely stripping the cyst Figs. , 3, 4, 5