During pregnancy, a variety of factors can influence fetal growth and development. Intrauterine growth may impact on later life and health. Neonatal body composition may be a more sensitive marker for the intrauterine environment than established anthropometric parameters at birth.To study neonatal body composition determined by air displacement plethysmography in healthy, term singletons as national reference data, and to establish factors impacting on neonatal body composition in this population. This prospective cross-sectional observational study included 271 healthy, full-term, singletons born between June 2014 and July 2015. Body composition was measured within 96 h of birth using air displacement plethysmography.Median  fat mass / total body mass (BF%) in German singletons was 10.8% (7.7–13.4) and fat free mass (FFM) 2843 g (2606–3099). Female infants had significantly increased BF% compared to male infants (11.2% (8.7–14.0) vs. 9.6% (7.2–12.1)). On multiple regression analysis, BF% and fat mass increased with female gender, maternal pre-pregnancy body mass index, non-smoking mother and parity, whereas FFM increased with male gender and increasing gestational age at birth. Gestational weight gain category, birth mode, and postnatal age at measurement were not associated with BF%, FFM or fat mass.We generated BF% and FFM centiles for healthy, term, singletons born in Germany; these are similar to those found in other European countries. Infant body composition at birth was associated with modifiable , and given factors . The incidence of obesity in children is rising worldwide. Currently, 17.0% of children in the United States are obese and the prevalence of extreme obesity is 5.8% . ObesityEpidemiological studies suggest that inadequate intrauterine supply of nutrients may impact on metabolic health in adulthood , 8. MostThere are different methods to determine neonatal body composition . For about 15 years, air displacement plethysmography (ADP) has been available as a method for rapid, non-invasive, pain-free determination of body composition at relative low cost, providing immediate results without ionizing radiation; thereby making body composition measurements in healthy children acceptable for parents and ethics committees. It has been shown that these measurements are highly reproducible and accurate and therefore suitable even for large epidemiological studies , 12. ADPReference data for healthy newborns are important as a basis to identify deviations in body composition from the reference standard in special patient groups  and to plan intervention studies aiming at improving modifiable pre- and postnatal factors affecting long-term health. For example, the aim of nutritional care of preterm infants is to achieve similar growth as in utero. Due to improvements in nutritional care of preterm infants, weight-gain as in utero is now often achieved , but bodDifferences in body composition between populations of different ethnic and/or socioeconomic background have been reported in adults and children , 16 and We aimed to generate reference data for German Caucasian population for BF%, FM and FFM at birth in healthy, term, singleton infants and to investigate factors influencing body composition.This was a prospective, cross-sectional study in a convenience sample of healthy, singleton, term infants (≥37 0/7 weeks gestation) born between June 2014 and July 2015 at Tuebingen University Women’s and Children’s Hospital, Germany. Infants were recruited postnatally by the study team on the puerperal ward if they fulfilled inclusion criteria. Parents were approached preferentially on the day after birth, to enable recuperation from birth. The aim was to address as many parents as possible, restricted by limited availability of the study team. Infants with major congenital anomalies  or severe diseases  and those born to mothers with pre-gestational or gestational diabetes mellitus were excluded.2) was calculated as pre-pregnancy weight divided by height squared. The following BMI categories were used: underweight (< 18.5); normal weight (18.5–24.9); overweight (25.0–29.9) and obese (> 30) [Maternal pre-pregnancy body mass index (BMI)  recommendations concerning the recommended gestational weight gain for singleton pregnancies depending on the maternal pre-pregnancy BMI were used to classify weight gain during pregnancy: underweight mothers ; normal weight mothers (11.5–16.0 kg); overweight mothers (7.0–11.5 kg) and obese mothers (5.0–9.0 kg) [The Institutional Review Board approved the study protocol and written informed parental consent was obtained.Data were collected from maternal health passports and medical records of the mother and her newborn, and parents were asked to fill in a questionnaire. Medical data included age, pre-pregnancy body mass index (BMI), parity, gestational weight gain, smoking during pregnancy and antenatal medical history. Paternal data included age and BMI. Neonatal data included age, sex, birth weight, length and head circumference.The PeaPod Infant body composition system  is an air displacement plethysmograph and can determine the body composition for infants between 1 and 8 kg body mass. Neonatal anthropometric measures and body composition were determined within 96 h of birth. After weighing, the naked infant was placed in the heated measuring chamber to determine his/her volume. The determination of body volume takes 2 min. BF%, FM, and FFM were calculated by the system as previously described , 19. Bodhttp://www.healthforallchildren.com/?product=lmsgrowth). The reference population was the British 1990 growth reference  if normally distributed, or as median and interquartile range if not. In case that within a table a minority of parameters were normally distributed, the data was nevertheless presented as median (Interquartile range) to improve clarity of presentation. Between group comparisons were performed using a two-sided t-test or ANOVA and post hoc Tukey’s multiple comparison test for normally distributed variables or a Wilcoxon test in non-normally distributed data and a Fisher’s exact test in categorical outcomes. Correlation between normally distributed continuous variables was assessed by linear regression and Pearson correlation coefficient. Associations of potential explanatory variables  with body composition parameters were assessed by multiple linear regression analysis with  backward elimination. Shapiro-Wilk test was used for assessment of normal distribution of data (before ANOVA and t-test) and residuals . Analyses were performed with GraphPad Prism® 8.1.0  and the level of significance was There were 3170 deliveries during the 1-year study period at Tuebingen University Women’s Hospital, Germany; 2649/3170 (83.6%) infants were born with gestational age > 37 weeks, 8/2649 (0.3%) of these died soon after birth or were stillbirths, 80/2649 (3%) infants were twins, 49/2649 (1.9%) infants had severe congenital anomalies and 16/2649 (0.6%) had severe diseases. Of the remaining 2496 singleton, healthy, term infants, 901 (36.1%) families were approached by the study team and 498/901 (55.2%) of these agreed to participate. Thus, 20.0% of eligible infants could be recruited.n = 20), use of a pacifier or blanket in the test chamber (n = 14), or discontinuation of measurement because of agitation or crying (n = 3), gestational age at birth < 37 weeks (n = 7) or twin pregnancy (n = 1)).In 133 of 498 infants recruited, body composition was not determined due to scheduling difficulties because of early hospital discharge (within 48 h of birth) and unavailability of study personnel on sporadic days. Forty-nine infants were excluded because of a maternal history of pre-gestational or gestational diabetes mellitus, and 45 were excluded for other reasons .Complete body composition measurements were available for 271 term infants  BF% in our population was 10.8% (7.7–13.4) and FFM was 2843 g (2606–3099). For a detailed description of the distribution of the parameters of body composition see Table  BF% in oGirls had higher BF% and tended to have higher FM than boys, whereas birth weight and FFM was higher in boys Table .Table 42 = 0.17, p < 0.0001), FM , and FFM  increased with increasing gestational age on linear regression. For comparisons between subgroups of different gestational ages see also Table Birth weight (r201 of 271 74.2%) infants were measured between 24 h and 72 h after birth with a median (Q1-Q3) postnatal age of 42 h (29–56). Day of measurement was not associated with changes in BF%, FM or FFM  and with each maternal BMI point, the offspring’s BF% increased by 0.2%. There was no correlation between absolute maternal weight gain in kg during pregnancy and body composition in linear regression. When the observed gestational weight gain was classified according to the 2009 IOM recommendations taking the pre-pregnancy BMI into account, 20.3% of participants’ mothers gained insufficient weight overall, 36.9% gained adequate weight, and 42.8% gained excessive weight. Excessive weight gain was associated with increased birth weight, BF%, FFM and FM  with a history of maternal smoking during pregnancy. Smoking during pregnancy was associated with lower birth weight, lower birth weight SDS (− 0.6 (− 1.5- -0.2) vs. 0.0 (− 0.6–0.6)), lower FM and trends towards lower BF% and FFM Table .BF% and FM tended to be higher with higher parity Table : first-bMultiple linear regression analyses indicated that gender, parity, pre-pregnancy BMI, and smoking were associated with BF% and FM  whereas FFM was associated with gender and gestational age only Table . In thisThe aims of this cross-sectional observational study were to establish reference data for body composition by air displacement plethysmography in healthy term singletons for Germany and to investigate various factors potentially influencing body composition.The infants studied here showed values for median BF% (10.8%) similar to those from other European countries such as Portugal (11.3%) , the NetIn our study population a significantly higher median BF% (11.2% vs. 9.6%) and lower median FFM (2786 g vs. 2977 g) were found in female infants compared to male infants, and gender was significantly associated with BF%, FM and FFM on multiple regression analysis. Consistent with this, the SCOPE pregnancy study, a large population-based study in Ireland, including 786 term infants confirmed differences in body composition between female and male neonates with girls having a higher BF% and lower FFM . In a crBesides ethnic factors and gender, increasing gestational age is associated with differences in body composition: Hawkes et al. described a significant and linear increase in BF% with increasing gestational age , but thiIn our study with measurement of body composition within the first 96 h, postnatal age at measurement was not associated with BF%, FM or FFM. In agreement with our results, there was no association of BF% with postnatal age in the SCOPE study , whereasBF% and FM were significantly associated with pre-pregnancy maternal BMI. Our results are consistent with previous studies using ADP. In a large pre-birth cohort study, the Healthy Start study , Starling et al. showed positive and independent associations of pre-pregnancy BMI with neonatal adiposity measures . PereiraIn our study, there was no correlation between absolute maternal gestational weight gain and body composition but a positive association between excessive weight gain by IOM recommendations and neonatal BF%. Additionally, inappropriately low weight gain was associated with decreased FFM in this German cohort. An altered body composition in infants born to mothers with excessive gestational weight gain was also found in the Healthy Start study : gestatiThe association of body composition parameters with pre-pregnancy BMI and gestational weight gain category reflect the impact of intrauterine environment on offspring obesity risk.There is evidence that obese infants and children are more likely to become obese adults, with an increased risk of characteristic complications (e.g. diabetes and cardiovascular disease) and increased mortality –6. Body n = 8) of mothers reported smoking during pregnancy, nevertheless multivariate analyses showed a significant association with lower BF% and FM, the magnitude of the effect of smoking exceeding that of gender. The apparent lack of impact on FFM may be due to the small numbers. The KiGGS Wave 2 study reported that 10.9% of German mothers smoked during pregnancy and that the proportion of pregnant women who smoke has declined over the last two decades [In this study, only 3%  BF% of 10.8% (7.7–13.4%) as reference for contemporary neonates for a German Caucasian population. These data are thus similar to those reported for other European countries. Factors associated with body composition in this study were gender, parity, smoking during pregnancy, pre-pregnancy BMI, gestational age at birth and gestational weight gain category, this information will be helpful for the design of future studies.


Background: Although most preterm infants breathe at birth, their respiratory drive is weak and supplemental oxygen is often needed to overcome hypoxia. This could in turn lead to hyperoxia. To reduce the risk of hyperoxia, currently an initial low oxygen concentration (21–30%) is recommended during stabilization at birth, accepting the risk of a hypoxic period. However, hypoxia inhibits respiratory drive in preterm infants. Starting with a higher level of oxygen could lead to a shorter duration of hypoxia by stimulating breathing effort of preterm infants, and combined with subsequent titration based on oxygen saturation, prolonged hyperoxia might be prevented.Study design: This multi-center randomized controlled trial will include 50 infants with a gestational age between 24 and 30 weeks. Eligible infants will be randomized to stabilization with an initial FiO2 of either 1.0 or 0.3 at birth. Hereafter, FiO2 will be titrated based on the oxygen saturation target range. In both groups, all other interventions during stabilization and thereafter will be similar. The primary outcome is respiratory effort in the first 5 min after birth expressed as average minute volume/kg. Secondary outcomes include inspired tidal volumes/kg, rate of rise to maximum tidal volume/kg, percentage of recruitment breaths with tidal volumes above 8 mL/kg, duration of hypoxia and hyperoxia and plasma levels of markers of oxidative stress (8-iso-prostaglandin F2α).Discussion: Current resuscitation guidelines recommend oxygen titration if infants fail to achieve the 25th percentile of the SpO2 reference ranges. It has become clear that, using this approach, most preterm infants are at risk for hypoxia in the first 5 min after birth, which could suppress the breathing effort. In addition, for compromised preterm infants who need respiratory support at birth, higher SpO2 reference ranges in the first minutes after birth might be needed to prevent prolonged hypoxia. Enhancing breathing effort by achieving an adequate level of oxygenation could potentially lead to a lower incidence of intubation and mechanical ventilation in the delivery room, contributing to a lower risk on lung injury in high-risk preterm infants. Measuring 8-iso-prostaglandin F2α could lead to a reflection of the true amount of oxygen exposure in both study groups. Although most preterm infants breathe at birth, their respiratory drive is weak and often insufficient to aerate their lungs and establish gas exchange –3. Histo2) during stabilization has been shown to be difficult , oxygen saturation (SpO2) and heart rates increased more slowly than the SpO2 references outlined in the international guidelines, despite the use of respiratory support. This indicates that when starting respiratory support with lower oxygen concentrations preterm infants are at a higher risk for hypoxia  resuscitated with an initial FiO2 of 1.0 . Further hypoxia . Indeed, hypoxia  and Whit hypoxia .2 level, in all studies most infants ended up with a FiO2 between 0.4 and 0.6 after 5 min to reach SpO2 target ranges. Currently, to prevent hyperoxia and achieve normoxia in newborns, international guidelines recommend to start stabilization after birth in preterm infants with low FiO2 (0.21–0.3)  , after wt ranges , 17, 27.t ranges .2 of 1.0, infants will reach SpO2 target ranges more rapidly, leading to less hypoxia and an increase in respiratory effort. With careful titration of FiO2 based on SpO2 target ranges, we hypothesize that the occurrence of hyperoxia will be minimized.We hypothesize that with an initial FiO2 of 1.0 compared to standard management of an initial FiO2 of 0.3 on breathing effort of preterm infants immediately after birth.The aim of this current study is to determine the effect of an initial FiO+0 and 29+6 weeks of gestation will be randomized to either start stabilization with an initial FiO2 of 1.0 or 0.3.The design of this multi-center study is a randomized clinical trial. Eligible infants between 24+0 and 29+6 weeks, born at the LUMC and the AMC will be eligible for inclusion. Infants with a congenital abnormality or condition that might have an adverse effect on breathing or ventilation will be excluded, such as congenital diaphragmatic hernia, trachea-esophageal fistula or cyanotic heart disease.The study will take place at the NICU of the Leiden University Medical Center (LUMC) and the NICU of the Amsterdam UMC (AUMC), both located in the Netherlands. Infants with a gestational age between 24The primary outcome is respiratory effort, expressed as average minute volume normalized for body weight in the first 5 min after birth.Secondary outcomes include;- Average tidal volumes in the first 5 min after birth- Average rate of rise to maximum tidal volumes in the first 5 min after birth- Percentage of recruitment breaths with tidal volumes above 8 mL in the first 5 min after birth- Duration of hypoxia  in the first 10 min after birth- Duration of hyperoxia (defined as oxygen saturation > 95%) in the first 10 min after birth- Changes in diaphragmatic activity in the first 5 min after birth- Incidence of intubation in the delivery room- Concentration of 8-iso-prostaglandin F2α , an oxidVolumes measured will be normalized for body weight. In addition, the interaction between those primary and secondary outcomes and the time after birth will be assessed.Other outcomes that will be collected include;- Percentage of time of facemask ventilation applied during stabilization at birth- Average oxygen saturation and heart rate in the first 5 min after birthper minute in the first 5 min after birth- Oxygen saturation and heart rate - Percentage of spontaneous breaths with tidal volumes above 4 mL/kg- Percentage of time that oxygen saturation is within 90–95% range in the first 5 min after birth- Percentage of time that oxygen saturation is within 90–95% range between 5 and 10 min after birth- Percentage of time that oxygen saturation is within target ranges defined by Dawson et al.  in the fIn addition, we will collect basic characteristics such as gestational age, birth weight, mode of delivery, incidence of complications in pregnancy, use of antenatal corticosteroids, maternal medication use, and Apgar score 1 and 5 min after birth.This study is not designed to demonstrate any differences in clinical short- and long-term outcomes. However, we will record study parameters focusing on short-term clinical outcomes, such as incidence of intubation in the delivery room and in the first 24 h after birth, pneumothorax, intraventricular hemorrhages (IVHs) grade 3 or more and neonatal death.2 of 1.0 compared to 0.3. In a recent study, the average minute volume of preterm infants over the first 100 breaths was 150 ± 70 mL/kg/min . Other parameters of respiratory effort increased as well: average expired tidal volumes (Vte) increased by 37% (from 4.9 to 6.7 mL/kg), and rate of rise to maximum tidal volumes increased by 32% (13.8 to 18.2 mL/kg/s)  in respiratory drive than demonstrated in the study of Van Vonderen et al.  sizes. Concealment of allocation will be ensured by using the randomization process of Castor EDC , an electronic data capture system. The neonatal fellow or neonatal consultant in charge of the procedure will randomize using Castor EDC.Allocation will be stratified by gestational age. Infants of 242 of either 1.0 or 0.3, after which FiO2 will be titrated following target SpO2 recommended in our local guidelines on stabilization after birth, which are based on the target ranges described by Dawson et al.  or a Giraffe Star System . Respiratory function monitoring at birth will be recorded during stabilization in the first 10 min of life, which is standard practice in our units. The respiratory function monitor will be used to measure inflation pressures, flow and tidal volumes. This can be used to measure average minute volume, average rate of rise to maximum tidal volumes, and the total time of ventilation given. It uses a small variable resistance anemometer to measure gas flow in and out of a facemask or endotracheal tube. This signal is automatically integrated to provide inspired and expired tidal volumes. The difference equals the leak from the facemask or endotracheal tube.In both groups, standard care will be provided in the delivery room and NICU and local guidelines on stabilization after birth will be followed, using a NeopuffBreathing effort will also be measured by activity of the diaphragm using electromyography (EMG) with the Dipha-16 . The Dipha-16 is used to record spontaneous breathing (EMG of diaphragm) and heart rate (ECG). Both parameters are used for audit of stabilization, which is standard care in the NICU of the LUMC. Three EMG electrodes connected to the Dipha amplifier (similar to ECG electrodes) will be placed. Two are placed bilateral at costo-abdominal margin in the nipple line, and the common electrode will be placed on one of the legs.2 and heart rate a Masimo SET pulse oximeter probe  or NellcorTM pulse oximeter probe  is placed around the right wrist or hand of the infant. The FiO2 is measured using a portable oxygen analyzer AX300-I , and the airway pressures are registered by a variable orifice flow sensor  connected to the facemask measuring the flow in and out the infant. The signals are digitized at 200 Hz using the NewLifeBox-R physiological recording system  and all signals are recorded by the NewLifeBox Neo-RSD computer system  supported by Polybench physiological software . Pulmochart software  is used to analyze primary and secondary outcomes.To record SpOChemical oxidation products of polyunsaturated fatty acids, formed by peroxidation initiated by free radicals, are shown to be the best biomarkers to assess oxidative stress in humans. Of oxidized fatty acids, the best indicator for oxidative stress is 8-iso-prostaglandin F2α . Cord blMeasurement time points:- Cord blood will be drawn directly after birth as a baseline value.- At ~1 h after birth, a blood sample will be taken together with the sample normally taken for measuring standard analyses; this ensures that the infant is not exposed to an extra venepuncture or capillary puncture. The total amount of inspired oxygen given from birth will be recorded from birth until the sample is taken.- At 24 h after birth, again together with measurement of standard blood analyses.For every sample, a total of 300 μL blood will be used to measure 8-iso-PGF2α. A preterm infant of 500 gram has ~40 mL of total blood volume, whereas an infant of 1,000 gram has ~80 mL. As only 2 samples will be taken from the preterm infants, only 600 μL of blood will be withdrawn, which accounts for 0.75–1.5% of total blood volume.2 setting, standard care is provided in both groups. Respiratory data is collected as standard clinical care. As the starting FiO2 is randomly assigned, we will ask for parental consent. Antenatal consent will be obtained before birth if the mother is hospitalized in the LUMC or AUMC with imminent preterm labor before 30 weeks of gestation. Whenever antenatal consent is obtained, but the mother passes the term of 30 weeks of gestation, the infant will not be included in this study.In addition to the initial FiOIn case the condition of the mother and/or fetus suddenly deteriorates, resulting in an emergency situation with immediate delivery, it may not be possible to ask for antenatal consent from either parent. Also, when the mother arrives in hospital in full labor and tocolysis is not an option, it may not be possible, or appropriate to ask for antenatal consent. Emergency situations after birth cannot always be predicted and the neonatologist may have to act upon the infant's interest without being able to discuss the treatment plan with the parents, let alone ask for study consent. The worsening of the condition of these infants, who are often also not pre-treated with antenatal steroids, will increase the chances for respiratory insufficiency after birth and the need for respiratory support in the delivery room. Exclusion of this group of infants will therefore cause an important selection bias.For the infants born in the following situations, retrospective consent will be used:In case of an emergency situation , when there is not enough time for antenatal consent.When obtaining antenatal consent is inappropriate . In this case there is enough time, though trying to obtain informed consent in this situation would be inappropriate.This study fulfills the criteria for a waiver of informed consent in Europe , Canada  and the United States (USA Food and Drug Administration 1999) and Australian NHMRC guidelines for studies in emergency medicine . For these infants, retrospective consent will be sought from the parents to use data obtained. For this important study to be feasible, and enroll a representative sample, a waiver of informed consent is appropriate when consent cannot be obtained prior to delivery.The parents will be informed as soon as possible after the study and asked to consent to data being collected on their infant. There are many precedents of randomized trials conducted with such a consent waiver in emergency situations in neonatal stabilization after birth and adult resuscitation.While undertaking research studies without prior informed consent needs careful consideration, studies of emergency medical interventions, such as neonatal stabilization after birth, are exceptionally difficult if prior consent is required. Arguably, it is more unethical to use treatments that have not been studied than to evaluate them in a carefully controlled manner , 33.Parents can withdraw their consent at any time for any reason if they wish to do so without any consequences. When parents want to withdraw their infant for any reason, this infant will be handled as drop-out. As we included an additional 10% of infants in the sample size calculation, this will not influence the power calculation.2 setting of 1.0. All adverse events observed by the investigator or the staff will be recorded and noted to the medical ethical committee of the LUMC.Adverse events are defined as any undesirable experience occurring to a subject during the study, considered related to the initial FiO2 of 1.0 vs. 0.3) as the intervention will be performed only at the start of stabilization, after which in both groups titration will be performed. In light of the above, immediate and individual reporting of all these conditions related complications will not enhance the safety of the study, and therefore these complications will be handled as adverse events.This study population (preterm infants) has a high risk of serious complications, which are inherent to their vulnerable condition and unrelated to the intervention that is under evaluation in this trial. These complications , retinopathy of prematurity (ROP), IVHs grade 3 or more, cystic periventricular leukomalacia (PVL), necrotizing enterocolitis (NEC) grade 2 or more, death) are not expected to be related to the intervention (starting stabilization after birth with a FiOData will be presented as median [interquartile range (IQR)], mean ± standard deviation (SD) or number (percentage) where appropriate. Data will be checked for meeting the assumption of normality first by making a frequency distribution per minute. Outliers will be identified and handled according to the data . Intention-to-treat will be employed. Statistical analysis will be performed with SPSS software version 23.0 .t-test if the data are found to be parametric, or with Mann-Whitney u-test when the data are non-parametric.Differences in average respiratory minute volume in the first 5 min after birth between the study groups will be tested with a Student's - The interaction of respiratory minute volume, tidal volumes, rate of rise to maximum tidal volumes and diaphragm activity in the first 5 min after birth with time will be analyzed with a linear mixed model, in which both study group (treatment) and time will be taken in consideration.t-test if the data is found to be parametric, or with Mann-Whitney u-test when the data is non-parametric.- Differences in average tidal volumes, average rate of rise to maximum tidal volumes, and percentage of recruitment breaths will be tested with a Student's 2 0.3 vs. 1.0) and time will be taken in consideration.- The concentration of 8-iso-prostaglandin F2α will be analyzed with a linear mixed model, in which both study group . The study protocol is approved by the ethical committee of the LUMC and AUMC.Results of the study will contribute to a manuscript, which will be submitted for publication to a peer-reviewed international medical journal, and will be presented at national and international conferences.2 target ranges  or when obtaining antenatal consent was inappropriate , consent was asked retrospectively. This study was registered in JD, MG, EM, GH, and AtP conceptualized the study. JD wrote the first draft of the manuscript. JD, SH, MG, EM, GH, WO, AvK, and AtP contributed to interpretation of the study protocol in relation to other studies performed on this subject (discussion). All the authors critically reviewed and contributed to the final draft 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.


There is a paucity of community-based data regarding the prevalence and impact of gout flares as these may often be self-managed. The aim of this study was to determine the prevalence of self-reported gout and gout flares, the use of urate-lowering therapy (ULT), and the association of gout flares with health-related quality of life (HRQoL) in a large community sample. Covariate associations with flare frequency and allopurinol use were also examined.n = 2778) were included in the analysis.The South Australian Health Omnibus Survey is an annual, face-to-face population-based survey. Data collected in the 2017 survey included self-reported medically diagnosed gout, allopurinol use , and gout attacks (flares) in the last 12 months, in addition to sociodemographic variables and health-related quality of life . Data were weighted to the Australian Bureau of Statistics 2016 census data to reflect the South Australian population. Participants 25 years and over (p < 0.05). The frequency of gout flares was associated with a lower physical HRQoL (p = 0.012). Current allopurinol use was reported by 51% of participants with frequent gout flares.The prevalence of gout was 6.5% . Amongst participants with gout, 37.1%  reported currently using allopurinol, while 23.2%  reported prior use (38% discontinuation rate). Frequent flares (≥ 2 in the last year) were reported by 25% of participants with gout and were more likely with younger age, higher body mass index, and current allopurinol use  contains supplementary material, which is available to authorized users. Gout is the most common inflammatory arthritis. The reported prevalence of gout is highly variable across the world, ranging from 0.1% to approximately 10%, with prevalence estimates greater than 1% in most developed countries , 3.9% inCurrent guidelines from the American College of Rheumatology (ACR) and the European League Against Rheumatism (EULAR) advise that long-term urate-lowering therapy (ULT), with the aim of maintaining serum urate levels  , 11, is Nonetheless, studies suggest a poor adherence to guidelines –16. ReasThere is a paucity of community-based data regarding gout flares. One internet-based case-crossover US-based study found that 53% of enrolled participants did not consult a health care physician during an acute gout flare , suggestThe aim of this study was to identify the prevalence of self-reported gout flare and the frequency of allopurinol use in a representative community-based survey. In addition, sociodemographic and clinical covariates associated with flare frequency and associations with comorbidities and HRQoL were sought.Data were obtained from the 2017 South Australian Health Omnibus Survey (HOS). The HOS is an annual, population survey, conducted by face-to-face interviews of approximately 3000 people aged 15 years and over, that obtains cross-sectional representative information on health, well-being, and related issues amongst the South Australian population living in metropolitan and rural areas. HOS has been designed to meet the highest standards of population survey methodology and is a clustered, multi-stage, systematic, self-weighting area sample , 24.The 2017 HOS survey consisted of data from 2977 interviews from 5300 selected households (participation rate 65.3%) and was conducted between September and December 2017.. A recent US study reported little change to gout therapy since the introduction of febuxostat to the market (prescribed to only 3% of gout study population) [Within the survey interview, three gout-related questions were asked, “Have you even been told by a doctor that you have gout?” with the response options of “Yes,” “No,” or “Do not know/refused.” To determine allopurinol use, the respondents were asked “Do you currently take/have you taken allopurinol for gout?” with the response options of “No, never taken” (never), “No, previously taken” (prior), or “Yes, still taking” (current), with a list of the current brand names of allopurinol available to the interviewees. Febuxostat was not included because it only became available on the Australian Pharmaceutical Benefits Scheme in 2015 as a second-line option if allopurinol was contraindicated or not tolerated . A recenulation) .To determine frequency of flares, respondents were asked, “If you have gout, how many gout attacks have you had over the last 12 months?” with the response options comprising of “None,” “One,” “Two,” “Three,” “Four,” or “Five or more.”Sociodemographic data collected included age, gender, and socioeconomic status (SES). SES was determined using the Index of Relative Socioeconomic Advantage and Disadvantage (IRSAD) , which iBody mass index (BMI) was based on self-reported height and weight, calculated according to standard formula, and was classified according to World Health Organization (WHO) criteria . Informat-scores with a mean of 50 and a standard deviation of 10 [HRQoL was measured by the SF-12 v1 (US version) with Physical Component Scores (PCS) and Mental Component Scores (MCS) computed as norm-based on of 10 .Analyses were performed using Stata v15.1 , and all tabulations, descriptive statistics, and regression models utilized appropriate survey weights. Only participants aged 25 and over were included in the analysis because gout is a disease of older adults. Prior to analysis, flares were grouped into three categories: none, 1, and ≥ 2 because, according to the ACR, ULT is indicated in patients with two or more gout flares/year .p values were reported.Data were weighted by the inverse of the individual’s probability of selection, as well as the response rate in metropolitan and country regions and then re-weighted to benchmarks derived from the Australian Bureau of Statistics 2016 Census data (age and sex). These person weights adjust the data to better align each case  with the age, gender, and geographic location distribution in the total South Australian population. Survey-weighted logistic , multinomial logistic , and linear (HRQoL) regression models were used to analyze relationships with relevant predictor variables. All models included the sociodemographic variables of age, gender, SES (IRSAD score), and BMI, with both linear and quadratic regression terms to allow for non-linearity in any relationship with the response variable. To enable interpretation of regression models, Stata post estimation commands were used to express results for each outcome as adjusted population-weighted marginal proportions/probabilities , or predicted means (for linear regression), for different levels of each predictor variable, and to determine the effect size  averaged over other covariates. For multinomial outcomes , Helmert contrasts of the outcomes were used to define meaningful comparisons, and joint Of the 2977 interviews conducted, 2778 participants, aged 25 years and over, were included in the analysis, with 71.4%  born in Australia and 10.0%  born in the UK. The estimated gout prevalence was 6.5% .p < 0.001), were older (p < 0.001), have a higher BMI (p < 0.001), and have a lower SES , BMI, and comorbidities between the general South Australian population and group with gout are presented in Table Current allopurinol was reported by 37.1% of the participants with gout, prior use by 23.2% (discontinuation rate of 38%), and 39.7% reported never using allopurinol (Table p = 0.009) and sex (p < 0.001) were associated with allopurinol use, whereas BMI (p = 0.63) and SES  were not. Older age was associated with a higher probability of allopurinol never-use , yet amongst allopurinol ever-users, a lower probability of discontinuation was observed . Females were also associated with a higher probability of allopurinol never-use , as well as a higher probability of discontinuation  amongst allopurinol ever-users.Sociodemographic and clinical covariates of allopurinol use were analyzed by multinomial logistic regression . There was no such trend for SF-12 MCS (p = 0.74).A majority of the participants with gout reported no flares in the last 12 months (58%); however, nearly 25% reported having two or more flares during this time (Table Only 51% 12.5/24.6) of participants with two or more flares/year were currently taking allopurinol (Table .5/24.6 op = 0.002), BMI (p = 0.005), and allopurinol use (p = 0.031) were the most important covariates associated with flares. Older age was most strongly associated with a decreased probability of flares . Although sex did not reach overall statistical significance (p = 0.10), possibly due to the relatively low proportion of females with gout, there was a trend that females were more likely to suffer from flares . A higher BMI was associated with an increased probability of flares , and within participants with flares, a higher probability of ≥ 2 flares . Current allopurinol use, within participants with flares, was most strongly associated with a higher probability of ≥ 2 flares .The relationships between allopurinol use, sociodemographic and clinical variables, and flares were analyzed by multinomial logistic regression  in the South Australian population aged 25 and over. This prevalence is comparable to previous population-based estimates in the South Australian population , 33, butThere are known difficulties in optimizing ULT for the management of gout. Current rheumatology guidelines recommend a treat-to-target approach, requiring regular serum urate monitoring and slow up-titration of the dose until target serum urate levels are achieved –12. FlarWe found that younger participants with gout had lower rates of allopurinol continuation and were more likely to have flares, findings that are comparable to those from two retrospective UK general practice database studies of people with incident gout , 41. ImpAlthough gout predominantly affects men, women were less likely to commence or adhere to ULT and experienced a greater number of flares in our study. Other studies have identified that women with gout have more severe disease with a greater burden of comorbid conditions  and poorThere are several limitations of this study. In addition to the use of self-reported, doctor-diagnosed gout, which has been validated for gout case definition , 38, flaThis is the first community-based study of gout flare, which utilized rigorous sampling methodology to ensure that the data was representative of the general population. We conclude that gout continues to be a prevalent and poorly managed disease, despite readily available treatment. A quarter of participants with gout reported frequent flares that were associated with reduced physical HRQoL. Current allopurinol use was reported by only 51% of participants with frequent gout flares, suggesting undertreated disease and suboptimal use of ULT.Additional file 1:Table S1. Sociodemographic variables as predictors of gout. Analysis was performed by survey-weighted logistic regression, and coefficients represent log-odds ratios. Table S2. Prevalence of gout by gender, age, and BMI. Table S3. Prevalence of gout % (95%CI) by age × gender. Table S4. Sociodemographic variables as predictors of allopurinol use (within gout participants). Analysis was performed by survey-weighted multinomial logistic regression, and coefficients represent log-odds ratios. Table S5. Sociodemographic variables and allopurinol use as predictors of the number of gout flares in the preceding year (within gout participants). Analysis was performed by survey-weighted multinomial logistic regression, and coefficients represent log-odds ratios. (PDF 139 kb)


Objective: The worldwide prevalence of anemia is ~24.8%. Iron deficiency anemia is common in children and women and associated with sensory, motor, cognitive, language, and socioemotional deficits. Therefore, detection and early intervention strategies for anemia in infants are urgently needed. To prevent the occurrence of iron deficiency anemia, we aimed to identify risk factors associated with anemia in infants.Methods: This investigation involved a cross-sectional study of 6-months-old infants discharged between April 2014 and September 2017 from Peking University First Hospital. We assessed birth information, maternal age, and maternal educational level as well as data on feeding style, complementary foods and primary caregivers. The infants were assessed with the Denver Developmental Screening Test (DDST).Results: A total of 1,127 6-months-old infants were enrolled at the hospital. We found that the prevalence of anemia among infants in Beijing was ~11.8%. Premature infants had a higher rate of anemia than full-term infants . Infants born in autumn or winter were at an elevated risk of developing anemia . Birth weight had no effect on the rate of anemia in infants . Infants who were exclusively breastfeeding had higher anemia rates than those who were fed formula . Infants whose caregivers added no complementary foods had higher anemia rates (24.7%) than those whose caregivers added more than two kinds of complementary food (8.2%). The type of caregiver had no effect on the anemia rate in infants .Conclusions: The following factors resulted in a higher prevalence of anemia in our study a gestational age at birth of <37 weeks, exclusive breastfeeding, a lack of supplementation with complementary foods and a spring birth date. No significant differences in DDST pass rates were evident between infants with and without anemia. Anemia is a common disease that affects ~1.6 billion people worldwide, especially infants and women. The World Health Organization (WHO) has estimated that the global prevalence of anemia to be ~24.8% . In rodeInfants aged 6–12 months are at an elevated risk of anemia because they are developing and growing rapidly and because the stored iron from the mother may be deficient. The addition of complementary food during this period is important. Complementary foods influence the overall nutritional status of the infant. The risk of iron deficiency increases in later infancy if infants are exclusively breastfeeding . MoreoveRapid economic development and the acceleration of industrialization in China have led to major changes in Chinese lifestyles, especially for new parents. As the capital city of China, Beijing is more representative of such changes than other cities. Infants in Beijing consume increasingly rich diets, and their developmental health is better now than in the past. As the education level of the mother increases, her knowledge of how to feed her children improves. In the present study, we aimed to investigate the factors currently associated with infant anemia in Beijing, with the goal of improving the health of these infants.This investigation involved a cross-sectional study conducted at Peking University First Hospital. The participants were enrolled between April 2014 and September 2017. The infants included in this study were 6 months old and did not have severe disease or any abnormality at birth. The exclusion criteria were as follows: younger or older than 6 months old, a history of asphyxia at birth, and a history of severe disease. Children who met the inclusion criteria and did not meet the exclusion criteria were enrolled in our study. For each infant, we collected data regarding sex , maternal education level , birth weight, birth season, caregivers , feeding style , and complementary food usage . We defined the four seasons as follows: winter , spring , summer , and autumn .Anemia is defined as an Hb concentration <110 g/L according to the WHO diagnostic criteria. Mild anemia is defined as an Hb concentration between 90 and 110 g/L, moderate anemia is defined as an Hb concentration between 60 and 90 g/L, and severe anemia is defined as an Hb concentration <60 g/L. Iron deficiency anemia is defined as a mean cell volume (MCV) <80 fl, mean cell hemoglobin (MCH) <27 pg, and mean corpuscular Hb concentration (MCHC) <310 g/L.The development of the infants' intelligence was assessed with the Denver Developmental Screening Test (DDST). The DDST was standardized for Chinese use in 1982 and has been utilized worldwide to assess the intelligence development of children aged 1 month to 6 years. The standardized DDST consists of 104 items and covers four areas of development: (a) personal/social, (b) fine motor/adaptive, (c) language, and (d) gross motor. In the present study, three trained professionals examined the children. The response options for the items were “passes,” “fails,” “refuses,” and “has not had the opportunity.” The results of the DDST could be normal (no delays), suspect (2 or more caution items and/or 1 or more delays), abnormal (2 or more delays) or untestable . The children with suspect or abnormal results were retested 2 or 3 weeks later.2 test and non-parametric tests were used to assess the differences in child development between the three groups. A P-value < 0.05 was considered statistically significant.The data were analyzed with SPSS 18.0. Numerical variables are presented as the mean ± standard deviation (SD) (birth weight). Enumeration data and ranked data are presented as percentages. ANOVA, the χThe study was carried out in accordance with recommendations of the Clinical Research Ethics Committee of Peking University First Hospital (Permit Number: 2017 [1375]). All parents provided written informed consent before the start of the study.A total of 1,127 infants  aged 6 months were included in this study. The average birth weights of the infants were 3.38 ± 0.42 kg for males and 3.26 ± 0.45 kg for females. The average birth lengths were 50.6 ± 2.1 cm for males and 50.2 ± 1.8 cm for females. The average Hb levels were 117.9 ± 8.5 g/L in males and 118.4 ± 7.8 g/L in females . The meaAs shown in 2 = 40.103, P < 0.001). The infants born in autumn or winter were at an increased risk of developing anemia . Birth weight had no effect on the rate of anemia in infants . Infants who were exclusively breastfeeding had higher anemia rates than infants who were fed formula . Infants whose caregivers added no complementary foods had higher anemia rates (24.7%) than infants whose caregivers added two or more types of complementary food (8.2%). The type of caregiver had no effect on infant anemia rates . P < 0.05).The factors that affected infant anemia are shown in 2 = 5.600, P = 0.051).As shown in The present study examined information associated with 1,127 6-months-old infants and revealed an anemia prevalence of 11.8%. In our study, the risk factors associated with anemia were gestational age at birth, birth season, feeding style and complementary food supplementation. No significant difference in the DDST pass rate was evident between infants with and without anemia.The anemia rate in our study was lower than the global anemia rate (24.8%) . The WHOIn our study, we found that premature infants had an increased risk of developing anemia at 6 months of age, which was consistent with other studies , 25. HalWe showed that infants born in spring had lower Hb levels than infants born in winter. A previous study also showed that the incidence of anemia in infants aged 5–7 months who were born in spring and summer was higher than that in infants aged 5–7 months who were born in autumn or winter . This diIn our study, we found that feeding style and complementary foods affected the prevalence of anemia in infants. Infants who were exclusively breastfeeding (17%) had a higher prevalence of anemia than infants whose diets were mixed or infants who were fed formula (4.4 and 6.1%). The formula consumed by the infants contained iron; therefore, the infants whose diets were mixed and the infants who were fed formula were unlikely to develop anemia. The addition of two or more types of complementary foods was associated with the lowest prevalence of anemia among the three groups (8.2 vs. 24.7% and 13.1%). The WHO recommends exclusive breastfeeding without the introduction of any nutritious complementary foods for the first 6 months . The conIn our study, we found that anemia was not associated with the DDST pass rates. This lack of association was probably because the duration of anemia in the infants was not sufficient to influence the DDST results. Iron requirements are most likely to exceed iron intake during the first 6–18 months of life because infants' growth and blood volume expansion proceed rapidly during this time . In our Our study showed that the gestational age at birth, birth season, feeding style and complementary food supplementation affected anemia in infants aged 6 months. Early detection is of utmost importance to prevent adverse outcomes caused by infant anemia. Next, the caregiver should add iron-containing complementary foods, such as liver paste, yolk and meat paste, at a suitable time, especially in infants whose gestational age at birth is <37 weeks. In addition, infants born during different seasons should be supplied with nutrition accordingly. For example, infants born during spring should be provided with more iron than those born during winter.Nevertheless, our study had several limitations. First, China is an expansive country that contains individuals of different ethnicities who inhabit different geographic locations and have various dietary traditions. These factors are probably associated with the prevalence of anemia. However, our study was limited to the population in Beijing. Further studies should focus on low-income and middle-income provinces in China. Second, we also lacked information regarding maternal anemia. Maternal anemia has been associated with infant anemia . In futuAnemia is a global public health problem that influences infant development, resulting in poor outcomes in adulthood. The risk factors identified in our study, such as a gestational age at birth of <37 weeks, exclusive breastfeeding, a lack of supplementation with complementary foods and a spring birth date, may be meaningful for the early detection of infant anemia and the prompt delivery of interventions.The raw data supporting the conclusions of this manuscript will be made available by the authors, without undue reservation, to any qualified researcher.The study was carried out in accordance with the recommendations of the Clinical Research Ethics Committee of Peking University First Hospital. All subjects gave written informed consent in accordance with the Declaration of Helsinki. The protocol was approved by the Ethics Committee of Peking University First Hospital, China.QL conducted the experiments, analyzed the data, wrote the manuscript, and approved the final version to be published. YH contributed to the conception and design of the experiment, acquired the data, revised the manuscript, approved the final version to be published, and agreed to be accountable for all aspects of the work. FL and WL contributed to the conception and design of the experiment, acquired the data, critically revised the manuscript, and approved the final version to be published. WS contributed to the conception and design of the experiment and approved the final version to be published.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.


The aim of our study was to explore the risk factors for very preterm (gestation under 32 weeks) and moderate preterm birth (gestation weeks 32-36 6/7) in singleton pregnancies in a national retrospective cohort study. We also wanted to establish whether IVF/ICSI is an independent risk factor for preterm birth after adjusting for already known confounders. We used data for 267 718 singleton births from 2002-2015 from the National Perinatal Information System of Slovenia, containing data on woman, pregnancy, birth, the postpartum period, and the neonate for each mother–infant pair. Mode of conception, maternal age, education, BMI, parity, smoking, history of cervical excision procedure, history of hysteroscopic resection of uterine septum, presence of other congenital uterine malformations, bleeding in pregnancy, preeclampsia or HELLP and maternal heart, and pulmonary or renal illness were included in the analyses. Unadjusted OR for very preterm birth after IVF-ICSI was 2.8 and for moderate preterm birth was 1.7. After adjusting for known confounders, the OR was still significantly elevated . Risk factors for very preterm birth with OR higher than 2.4 were history of cervical excision procedure, resection of uterine septum, operation or having other congenital uterine malformations, and bleeding in pregnancy. Risk factors for very preterm birth with OR between 1.4 and 2.1 were age >35 years, being underweight or obese, not having professional education, smoking, first birth, preeclampsia/HELLP, and IVF/ICSI. Risk factors for moderate preterm birth with OR higher than 2.4 were history of cold knife conization and other congenital uterine malformations. We found that even after adjustment, IVF/ICSI represents a single risk factor for early and late preterm birth even after adjustment with other risks such as maternal age, smoking, or a history of invasive procedures for either cervical intraepithelial neoplasia or infertility treatment. Assisted reproductive technology is now widely used for treatment of different female and male causes of infertility. More and more babies are born after IVF/ICSI procedures each year worldwide. The same is true for Slovenia, where up to 4% of babies are born after IVF yearly . IVF wasMultiple factors were found so far to be connected with preterm birth, the most prevalent being extremes of maternal age, low maternal BMI, maternal smoking, infections, history of cervical excision procedures, uterine anomalies, infertility treatment, and others , 11.Women conceiving after IVF/ICSI are a special population of pregnant women. Due to many years of infertility, they are older than women conceiving spontaneously. The infertility itself is a known risk factor for preterm birth , since dThe aim of our study was to explore the risk factors for very preterm birth, before 32 weeks of gestation, and moderate preterm birth, from 32 to 36 6/7 weeks of gestation, in singleton pregnancies in a large 14-year national study. We wanted also to further explore weather pregnancy after IVF/ICSI is an independent risk factor for very preterm birth and moderate preterm birth after adjusting for already known important confounders.We conducted a population-based retrospective cohort study using data from Medical Birth Registry, the National Perinatal Information System of Slovenia (NPIS). We followed the methods of Jančar et al. 2016 . NPIS coThis retrospective cohort did not need ethical approval according to Slovenian law .The study population consisted of Slovenian residents who gave birth to singletons from January 1, 2002, to December 31, 2015. In this period there were 282 517 births in Slovenia. After exclusion of foreigners, Slovenian residents had 281 358 births in this period. After exclusion of 9800 multiple gestations , 3 836 (1.4%) induced births and elective cesarean sections before 37 weeks of gestation due to maternal and fetal conditions, and 4 cases with gestational week at the time of birth not recorded, our final sample for the analysis consisted of 267 718 spontaneous births of singletons.To be able to analyze factors associated with spontaneous preterm birth, we excluded all induced births and elective cesarean sections before 37 weeks of gestation, which were carried out due to maternal or fetal illness or condition, such as preeclampsia, maternal chronical illness, intrauterine growth restriction, or other critical conditions. Only births with a spontaneous onset have been included.Outcome variables were spontaneous preterm birth before 32 weeks and spontaneous preterm birth between 32 and 36 6/7 weeks of gestation. Gestational age was determined according to the last menstrual cycle and first ultrasound in pregnancy by obstetrician or gynecologist, in agreement with the paediatrician assessment after birth. In cases of unclear gestational age, it was determined individually, considering the anamnesis of last menstrual period, including possible alterations such as irregular cycles, combined with the first available ultrasound estimation and findings of the pediatrician. The most probable gestational age was included in the system.2, 18.5–24.9 kg/m2, 25–29.9 kg/m2, and 30 kg/m2 and higher. Parity was categorized into three groups: first birth, second birth, and third or more. Reported smoking during pregnancy was categorized into two groups: no or yes. History of cervical excision procedure was categorized into three groups: no, history of cold-knife conization, history of newer excision procedures, and predominately large loop excision of transformation zone (LLETZ). History of hysteroscopic resection of uterine septum was categorized into two groups: no or yes. Presence of other congenital uterine malformation was categorized into two groups: no, when there was no anomaly, and yes, when it was present or when it had been surgically corrected before this pregnancy. Bleeding in pregnancy was categorized into two groups: no and yes, when there had been a history of bleeding anytime in this pregnancy. Preeclampsia or HELLP in this pregnancy was categorized into two groups: no or yes. A history of maternal heart and pulmonary or renal illness was categorized into two groups: no or yes.Along with mode of conception, IVF/ICSI or spontaneous, a total of 11 other covariates obtained from NPIS were included in the analyses: maternal age, maternal education, maternal BMI, parity, smoking during pregnancy, history of cervical excision procedure, history of hysteroscopic resection of uterine septum, presence of other congenital uterine malformations, bleeding in pregnancy, preeclampsia or HELLP in pregnancy and maternal heart, and pulmonary or renal illness. We selected those covariates, because they have been previously reported to affect the risk for preterm birth . Maternap) values. A p value of < 0.05 was considered statistically significant. For statistical calculations, we used IBM SPSS Statistics for Windows, Version 21.0. .Chi-square test was used for descriptive analysis. Logistic regression analyses were performed to estimate crude odds ratio (OR) and adjusted odds ratio (aOR) and their 95% confidence intervals (95% CI) with two-sided probability  singletons were conceived after IVF/ICSI. The percentage of pregnancies after IVF/ICSI rose constantly over the years: from 1.5% in the year 2002 to 3.9% in the year 2015. The characteristics of our study population according to distribution of covariates divided by mode of conception is shown in A total of 11 605 singleton births (4.3%) in our population were premature, before 37 weeks of gestation. The distributions of preterm births according to gestation and mode of conception are also shown in We have calculated the unadjusted odds ratios (OR) for very preterm birth, before 32 weeks of gestation and for moderate preterm birth, between 32 and 36 6/7 of gestation, for women, who conceived after IVF-ICSI, compared to women conceiving spontaneously. Unadjusted OR for very preterm birth in pregnancies after IVF-ICSI conception was 2.8, for moderately preterm birth 1.7. After adjusting for included known confounders, the OR remained statistically significantly elevated. The results are presented in We also prepared multivariate analysis of different factors contributing to premature birth in our study population taking into account twelve covariates: mode of conception, maternal age, maternal education, maternal BMI, parity, smoking during pregnancy, history of cervical excision procedure, history of hysteroscopic resection of uterine septum, presence or previous surgical correction of other congenital uterine malformation, bleeding in pregnancy, preeclampsia or HELLP in pregnancy and a history of maternal heart, and renal or pulmonary illness. The risk factors for very preterm birth, before 32 weeks, and moderate preterm birth, between 32 and 36 6/7, are presented in According to this multivariate analysis, there are some factors contributing to the risk for very preterm birth in our population. Those factors are age more than 35 years, being underweight or obese, not having any professional education, smoking during pregnancy, first birth, bleeding, preeclampsia or HELLP during this pregnancy, and pregnancy after IVF/ICSI, which all have OR between 1.4 and 2.1. Similarly, factors for moderate preterm birth between 32 and 36 6/7 having OR between 1.4 and 2.1 are age more than 40 years, being underweight, first birth, LLETZ, and bleeding in pregnancy.The most important risk factors for very preterm birth, which have OR higher than 2.4, are having a history of any cervical excision procedure, having previous hysteroscopic resection of uterine septum, having or being previously operated for other congenital uterine malformation, or bleeding any time in pregnancy. The most important risk factors for moderate preterm birth which have OR higher than 2.4, are having a history of any cervical excision procedure, having previous hysteroscopic resection of uterine septum, and having or being previously operated for other congenital uterine malformation.Our results showed that the risk of sPTB in singleton pregnancies after IVF/ICSI is significantly greater than that in spontaneously conceived singletons. Our findings are in agreement with the most recent meta-analysis of Cavoretto et al. . In thisAs presented in For women, who conceive after IVF/ICSI, it is significantly more likely that they are nulliparous and have significantly more cervical excision procedures, significantly more resections of uterine septum, and significantly more other congenital malformations than women who conceive spontaneously. Furthermore, they are more likely to bleed at any time in pregnancy and to have other complications such as preeclampsia and HELLP than women, who conceive spontaneously. Therefore we performed an adjusted OR analysis. In this analysis, the OR for very preterm and moderate preterm birth remained significantly higher, which implies that IVF/ICSI might be an independent risk factor for preterm birth. Underweight patients in our analysis have higher OR for preterm birth than overweight or obese patients, which has been described before . RegardiCold knife conisation and LLETZ were already shown to be important independent risk factors for premature birth in our recent population-based cohort study  and in sHysteroscopic resection of uterine septum was performed in 18.6% women (n = 1084) after IVF/ICSI and in only 3.5% of women (n = 9 111) conceiving spontaneously (p < 0.001). This is the consequence of diagnostic and operative interventions to find the exact cause of infertility before sending the infertile couple to IVF/ICSI. In Slovenia, we tend to have a holistic approach to infertility and we try to treat it in causative manner. Therefore, if we do not find a clear cause of infertility, such as severe teratozoospermia or blocked fallopian tubes, we perform diagnostic hysteroscopy and laparoscopy . If a utThose are the reasons for such a big discrepancy between the two groups of women. It is known that the presence of untreated uterine anomalies is connected to preterm birth . The hysDue to these findings, the procedures on cervix and uterus should be clearly indicated, especially in infertile population, since they might change integrity of the cervix in subsequent pregnancy.Bleeding any time in pregnancy was also more common in the group of women after IVF/ICSI. It is already known that bleeding and spontaneous abortion in first trimester occur more common after IVF/ICSI than after spontaneous conception. To some extent, this can be contributed to higher aneuploidy rate due to the advanced age of both partners, mostly female, but also male partner , 31. On Placenta previa and certain other placental abnormalities, followed by antepartum hemorrhage, generally occur more often in pregnancies after IVF and the causes are still obscure –36. OthePreeclampsia is also more common in pregnancies after IVF, mostly due to infertile population attributes, since after adjustment for the confounders the association was weak .This study confirms findings that the population of women, who conceive after IVF/ICSI, is different than the population of women, conceiving spontaneously, and they deserve a closer and more dedicated follow-up during their long wanted pregnancy.Furthermore, in this large cross-sectional study we proposed a combination of risk factors that define population of women with highest risk for spontaneous onset of preterm delivery.According to the findings, in IVF patients, the cervical neoplasia treatment methods and hysteroscopic operation techniques should have been analyzed further, to narrow groups with the highest preterm birth risk. The cervical ostium assessment prior to hysteroscopy could be of some importance.Besides, different IVF treatment methods should have been analysed, since the perinatal outcome is not the same in all cases: children, born after fresh ET, are at higher risk for low birthweight and premature delivery as children, born after frozen-thawed ET , 44.In our study no late abortions with stillborn fetuses under 500 g and before 22 gestational weeks were included. This data could possibly show another distribution of risk factors.There are still numerous known preterm birth risk factors, including previous preterm birth, asymptomatic bacteriuria, sexual intercourse, and psychiatric disorders, which have not been analyzed or added to our model and could add an insight to the etiology of preterm birth.Our study was designed as a population-based cohort study. The main advantage of a population-based study over 14 years is the large number of births available for analysis and inclusion of all population subgroups . In Slovenia, there is a universal access to assisted reproductive procedures, regardless of social status. The costs for six cycles of IVF for the first child and four cycles for the second or third child are covered by medical insurance for every infertile couple, where another treatment of infertility was not successful or not possible. This fact and total population inclusion diminishes the existence of selection bias in our study.We realize that by employing data from administrative sources one could question the quality of such data. As described, the data source is obligatory by law, is in use for more than 30 years in Slovenia, is predefined, has regular quality checks, and is made of data gathered in medical records which are produced by medical staff in maternity hospitals; this is why we consider our data to be of reasonably good quality.One of the weaknesses of our study that was already mentioned is the fact that no late abortions were included in the analysis.In this large cross-sectional national study we proposed a combination of risk factors that define population of women with higher risk for preterm birth. We found that even after adjustment IVF/ICSI represents a single risk factor for early and late spontaneous preterm birth even after adjustment with other risks such as maternal age, smoking, or a history of invasive procedures for either cervical intraepithelial neoplasia or infertility. Therefore, these women deserve a more close and dedicated follow-up during their pregnancy.


To construct birth weight charts for the Chongqing municipality, China and to identify whether differences in birth weight exist across urban/rural populations, thereby warranting separate charts.Secondary analysis of routinely collected data from 338,454 live infants between 2014 and 2017 in Chongqing municipality. Sex-specific birth weight-for-gestational age centiles were constructed by the lambda-mu-sigma method via the GAMLSS R-based package. This method remodels the skewed birth weight distribution to estimate a normal distribution, allowing any birth weight centile to be generated. A separate set of centiles were created, accounting for urban/rural differences in birth weight.n = 4176) of males and 2.26% (n = 3656) of females were classified as below the 2nd centile (expected percentage = 2.28%), 49.75% of males  and 50.73% of females  were classified as below the 50th centile (expected proportion = 50%) and 97.52% of males  and 97.48% of females  were classified as below the 98th centile (expected proportion = 97.72%). The overall estimated centiles of birth weight for rural infants were higher than the centiles for urban infants at the earlier gestational ages . However, this trend was reversed in infants born at term.The centiles performed well across all gestational ages. For example, 2.37% (We have constructed a readily utilizable set of birth weight references from a large representative sample of births in Chongqing. The method used to construct the references allows for the calculation of the exact centile for any infant delivered between 28 and 42 completed weeks, which was not possible with previous charts. Birth weight is a key measure of infant health, providing an indication of prenatal wellbeing, risk of postnatal complications and survival . The ideAn important consideration in the use of the reference charts is the population on which the charts were constructed. For example, ethnicity, period of data collection (secular changes), socioeconomic status and whether the sample adhered to a set of strict inclusion criteria (e.g. the WHO standard ) should Data were extracted from the Chongqing National Population-Based Birth Defects Surveillance System (CNPBDSS). The CNPBDSS is a subset of the National Population-Based Birth Defects Surveillance System (NPBDSS) which was established in 2006 by the Ministry of Health, China to record all livebirths, fetal deaths and stillbirths. The system covers 64 counties and districts under the central government. Further details have been described elsewhere , 12. GesWe collected the data of 348,454 single births from 62 hospitals in Chongqing. Informed consent for use of the data was obtained when the Birth Defects Register Forms were collected , 12.+ 0 and 42+ 6 weeks gestational age. Infants were not included if sex or birthweight was not recorded or undefined, or if the birthweight was considered as an outlier. An outlier birthweight was defined as being greater than four standard deviations from the median birthweight for their gestational age and sex.Analyses were restricted to live-born infants born between 28An a priori decision was taken to use the LMS  method to calculate the new birth weight centiles , thus as‘pb’ function was implemented, with gestational age on the raw scale (weeks). The ‘pb’ function automatically selects the edf which results in the best-fitting model, as determined, in this analysis, by the lowest value for the Bayesian Information Criterion (BIC). Accordingly, the resulted edf for the L, M and S parameters were 2.00, 8.52 and 5.93, respectively, for males and 3.57, 9.72 and 6.18 for females. Residual diagnostics (via checking de-trended Q-Q plot (‘worm plots’) parameters and the normality of residuals) were performed and suggested that the BCCG distribution provided a suitable fit to the data.The LMS method was implemented using the GAMLSS package in R version 3.2.3, as suggested by the World Health Organisation . This apThe smoothed values of L, M and S were then applied to transform the observed distribution of birthweight to an accepted Normal distribution. Using these LMS values, any birth weight centile could be outputted at any gestational age via the formula:za is the normal equivalent deviate corresponding to a given centile [where  centile . LMS valPost creating the centiles, we defined the percentage of births below the set of standard centile thresholds, both overall and by gestational age group: < 32 weeks ; 32–36 weeks  and > 36 weeks (Term). For example, if the derived centiles calculate accurately, 10% of birthweights should fall below the 10th centile.The decision to use the BCCG (LMS) distribution was motivated by a desire to be able to use the resulting charts to produce an exact centile or z-score for a given infant, which is not straightforward with other distributions  or Box-Cox T (BCT). However, as a sensitivity analysis, we utilised the ‘lms’ function in R which fits a number of distributions to the response variable (birth weight), in order to find the optimal distribution (lowest maximum likelihood). It was observed that the Box-Cox T distribution (BCT) provided the best fit to the data and we have reported the proportions of infants below the same set of centiles, as well as the weekly birth weights at these centiles, in the supplementary material.In order to determine whether the distribution of birth weight was different in infants born to mothers who delivered in hospitals within the Urban developed economic circle (urban) compared to infants born to mothers who delivered outside of this area , we repeated the above steps after incorporating a covariate for urbanicity, which resulted in the production of area-specific LMS parameters and birth weight centiles.We sought to compare the proposed centiles to those of the ‘national population-based references’ of Dai et al. (2014). Particularly, we calculated the absolute (gram) and percentage /proposed centile value) *100) differences between the birth weight values at the 10th, 50th and 90th centiles of our charts to those of the Dai et al. charts. This was performed at 28–42 completed weeks of gestation because this was the period of overlap between the two charts.n = 452) or being born outside of 28–42 completed weeks (n = 333), leaving a total of 338,454 birth weights included in the analysis. The median birth weight of the sample was 3280 g (interquartile range (IQR): 3000; 3550), with a median gestation of 39 completed weeks .A total of 339,239 singleton births occurred in the 62 hospitals sampled from the Chongqing municipality between 1st January, 2014 and 30th September, 2017. Of those, 785 singletons were excluded from analysis due to: an implausible birthweight ), 50th and 90th ) centiles are presented in Additional file n = 133,596) of the total sample. Urban and rural sex-specific centiles are shown in Fig. Urban births accounted for 39.47% . Newborns with suspected IUGR may benefit from carefully monitoring in the period subsequently following birth for conditions such as hypoglycaemia.The centiles were produced using the LMS method . Using tWe observed marked differences in the birth weight distribution of infants whose mothers delivered in ‘urban’ vs ‘rural’ areas. Specifically, in the preterm period (< 37 weeks) we observed markedly higher birth weights for a given centile in infants born in rural areas compared to urban areas. This pattern was reversed in term births, such that ‘urban’ infants had higher birth weights at equivalent centiles. The authors conclude that access to better antenatal care may explain this observation. We speculate that the main contributor to the differences between the two groups at the earlier gestational ages is due to selection bias created via the improved antenatal care experienced by those delivering in hospitals within the central Chongqing municipality. For example, an infant born at 28 weeks who is delivered in a hospital within the central Chongqing municipality with access to more advanced healthcare facilities, has an increased likelihood of survival than the same baby born in a rural area. As centiles are based on live births, this would permit this lower birth weight to contribute to the urban centiles, whereas only the heavier (and thus more likely to survive) rural babies contribute to the rural centiles, thus shifting the birth weight distribution upwards at these earlier gestational ages. We hypothesise that some of the urban-rural variation could also be the result of differences in the proportion of pregnancies dated via ultrasonography. It has been shown that birth weights derived from ultrasound dated pregnancies have a higher population mean and their curves show less flattening at term than birth weights from pregnancies dated using LMP . In addiCost is considered the main obstacle to the uptake of maternal care, particularly for poor rural households. Since the economic reforms introduced in the 1980s, which saw central government decentralise fiscal responsibility to local government and a concomitant decline in central government transfers into the health sector –21, healAt term, the direction of the urban – rural birth weight difference was reversed. We speculate that this is likely the result of better nutritional status of women in urban areas which may have resulted in increased birth weight.n = 112), we did not have enough data to estimate reliable centiles at these gestational weeks. As such, in those periods where there were less data, the performance of the centiles was poor. Despite the charts constructed assuming a BCT distribution  resulting in improved performance at the earlier weeks, the differences were generally small. Besides, antenatal characteristics of the mother are not known.Due to the low number of births in our cohort that occurred in the earlier gestational weeks  to provide a greater indication of the severity of abnormal fetal growth.Additional file 1: Table S1. L, M and S values to produce any specified centile by sex and completed week of gestation. Table S2. Expected and observed percentage (n) below selected centiles: very preterm, late-moderate preterm, term, all gestational ages (BCT distribution). Table S3. Weekly birth weights (g) at the 10th, 50th and 90th centile. Table S4. Weekly birth weights (g) at the 10th, 50th and 90th centile (BCT distribution). Table S5. Comparison of new centiles vs Dai et al. (2014)a centiles.