Introduction: Human exposure to parabens is very common in daily life, and prenatal exposure to these chemicals is associated with poor birth outcomes. Therefore, the aim of this study was to investigate the effect of glutathione S-transferase (GST) polymorphisms on the association between prenatal exposure to parabens and birth outcomes. Methods: We conducted a multivariate analysis involving 177 subjects to determine the association between paraben concentrations and birth outcomes in mothers with GST mu 1 (GSTM1) and GST theta 1 (GSTT1) polymorphisms from 2017 to 2019. Furthermore, we determined the interactive effect between paraben levels and GSTM1/GSTT1 polymorphisms using regression analysis, in addition to a generalized linear model after stratifying GSTM1/GSTT1 genotype into three categories. Results: Methyl and propyl paraben concentrations were significantly and positively associated with birth weight  = 22.460–210.590; propyl, β = 82.352, 95% CI = 9.147–155.557) in individuals with the GSTM1-null genotype. Moreover, the propyl paraben concentration was significantly associated with an increase in gestational age  in individuals with the GSTM1-null genotype. Conclusions: This study reported the association between prenatal paraben exposure and birth outcomes in individuals with GST polymorphisms. We found positive relationships of maternal exposure to methyl parabens with birth weight in both mothers with GSTM1 and GSTT1-null genotypes. Humans are affected by various environmental factors. Environmental phenols such as parabens, triclosan, benzophenone-3, and bisphenol A are used in many industrial products. The majority of phenols are found in products used daily such as antibacterial soap, deodorant, household cleaners, kitchenware, toys, cosmetics, sunscreen, food, plastic packaging, perfumes, and plastics.Parabens, esters of p-hydroxybenzoic acid, are one of the most encountered phenols in such products. They have chemical stability, as well as non-volatile and antibacterial properties, and thus they are used widely as preservatives in cosmetics, personal care products, food, and some medicines ,2. ParabBirth outcomes, including weight, height, and gestational age at delivery, are significant predictors of fetal and infant health . PretermSeveral studies have demonstrated a relationship between parabens and birth outcomes. Some studies have shown a positive relationship between prenatal exposure to butyl parabens and placental weight ,14. The In addition to environmental factors, genetics influence birth outcomes. Glutathione S-transferase (GST) enzymes, including GST theta 1 (GSTT1) and GST mu 1 (GSTM1), play significant roles in detoxification of environmental carcinogens ,19,20,21Thus, the objective of this study was to provide evidence for an association between prenatal exposure to parabens and birth outcomes in pregnant women with GST polymorphisms, as no study has investigated this to date.This study is part of the ongoing Mothers and Kids Environmental health (MAKE) trial, which was initiated in 2017. We recruited pregnant women (≥28 weeks of gestation) who can visit study site during pregnancy in metropolitan area in Korea. A face-to-face interview was conducted to collect information, including socioeconomic status, habits of plastic and cosmetic use, and exposure to parabens. Urine and serum samples were collected at the time of visit to measure biomarkers including genetic information. We also collected the most recent sonographic findings at the time of visit and followed them up until birth to obtain data on the outcome of the birth. The study protocol was approved by the National Cancer Center of the National Cancer Center of Korea , and we obtained written consent from all participants prior to participation in the study.By this time, a total of 291 pregnant women were enrolled in this study, but ultimately 243 women were visited. Six women with twin pregnancies and 2 women with no specific gravity due to insufficient urine specimens were excluded. A total of 40 women with a paraben level below the limit of detection (LOD) and 18 women with missing follow-up data on birth outcomes were excluded. Therefore, 177 participants were finally included in the analysis.v/v), and the total analysis time was 18 min. For precise reproducibility of the data, we found that the concentrations of methyl parabens evaluated were 10, 100, and 300 µg/L (coefficient of variation (CV) 2.2–7.1%). The CV range for ethyl parabens was 3.1–6.7%, with 1, 12, and 50 µg/L, while that for propyl parabens was 3.3–5.1%, with 1, 12, and 50 µg/L. In an accuracy analysis reflecting the total error of the analysis method, the concentrations of methyl parabens evaluated were 10, 100, and 300 µg/L, representing 100 ± 5.6% values; those for ethyl parabens were 1, 12, and 50 µg/L, representing 102.3 ± 5.6% values; and those for propyl parabens were 1, 12, and 50 µg/L, representing 98.4 ± 3.8% values. The LOD was defined as the minimum concentration of parabens that could be detected in the analyte.Spot urine samples were stored in 15 mL conical tubes and refrigerated immediately at 2–6 °C. The tubes were transported within 24 h and stored in an ultra-low temperature freezer at −20 °C. High-performance liquid chromatography mass spectrometry (Agilent 6490) was used for urine separation and quantitative analysis. We used the RP18-e (100–103 mm) separator to isolate the materials for analysis, at a flow rate of 0.45 mL/min and temperature of 40 °C. Water and acetonitrile were used as the mobile phase solvents at a ratio of 85:15  after determining the lowest concentration of the calibration curve standard solution present in the samples. In this analysis, the LOD values were 0.116, 0.105, and 0.085 μg/L for methyl, ethyl, and propyl parabens, respectively.Urine creatinine levels were often used to correct for dilution of urine, but creatinine levels were not stable due to rapid changes in renal creatinine clearance during pregnancy. Thus, we decided to use the urine-specific gravity, which does not depend on the factors that explain the urinary dilution. BPs concentrations were corrected for specific gravity using the formulas suggested by Cone et al. .(1)BSG=Genetic information was obtained from maternal blood samples, which were centrifuged in cryovials, divided into 1 mL aliquots, and stored at −80 °C. GST polymorphisms were verified by conducting multiplex polymerase chain reaction (PCR). Before PCR, 2 µL DNA was added to the reaction mixture, which was subsequently microcentrifuged. The PCR program consisted of 94 °C for 5 min (pre-denaturation step), 35 cycles of 94 °C for 30 s, 60 °C for 30 s, and 72 °C for 30 s, followed by 72 °C for 7 min (post-extension step) for the GSTM1 genotype, or 94 °C for 5 min (pre-denaturation step), 40 cycles at 94 °C for 40 s, 60 °C for 40 s, and 72 °C for 40 s, followed by 72 °C for 7 min (post-extension step) for the GSTT1 genotype. The amplified PCR products were subjected to agarose gel electrophoresis  and ethidium bromide staining (0.5 µg/mL) for 15 min and were observed under ultraviolet light. The GSTM1 and GSTT1 products were represented by bands at 215 and 480 bp, respectively. The internal control for GSTM1 and GSTT1 was β-globin, represented by a PCR product band at 268 bp.t-test and analysis of variance. Paraben concentrations were log10-transformed due to their right-skewed distributions. We examined the correlation between pairs of log10-transformed urinary parabens concentrations using Pearson correlation coefficients. The effects of confounding factors on urinary paraben concentrations were tested using the Wilcoxon’s rank sum and Kruskal-Willis tests, and are expressed as median values due to their non-normal distributions. Concentrations below the LOD are presented as the LOD divided by the square root of 2. Confounding factors that potentially influence the effects of paraben exposure on birth outcomes were obtained from the previous literature [2), past history of alcohol consumption (yes or no), past history of smoking (yes or no), sex of the newborn , gestational age at delivery (<37 or ≥37 weeks), birth weight (<2500 or ≥2500 g), and parity (0 or ≥1). We decided to include patients with any missing data on birth height in the statistical analysis due to our limited sample size.The general characteristics of the study subjects are expressed as numbers and percentages or as means ± standard deviation. The effects of the baseline characteristics on all birth outcomes were evaluated using Student’s terature ,16 and iTo examine interactive effect of the genotypes and paraben concentrations on birth outcomes, we included an interaction term between paraben concentrations and genotypes. After identifying the interaction effect, we performed stratified analysis by genotypes of GSTM1/GSTT1, and combined effects of GSTM1 and GSTT1 genotype were also assessed using regression analysis. To examine the effects of maternal genotype on birth outcomes, we performed regression analysis, with adjustments for covariates to prevent the effects of confounding factors. The interaction term between the genotypes and paraben levels was analyzed with relation to birth outcome.p-value < 0.05 was considered to indicate statistical significance in all statical model including interaction test.In addition, we used a generalized linear model to stratify the GSTM1/GSTT1 genotypes into the following three categories: present, null, and double null. All statistical analyses were conducted using SAS statistical software . A Study participant characteristics are shown in r = 0.65, p < 0.0001). When we analyzed the relationship between parabens according to GSTM1 and GSTT1 genotype . The median concentrations of parabens stratified by the participant characteristics are shown in p < 0.001).The parabens detected at a rate > 90% are shown in genotype , we alsoβ = 147.77, p = 0.031; propyl paraben: β = 114.725, p = 0.013). No significant associations between birth outcomes and other GSTM1/GSTT1 genotypes were found.The MAKE study analyzed 177 women during the third trimester of pregnancy between 2017 and 2019 in Korea. The aim of this study was to investigate the effects of GSTM1 and GSTT1 polymorphisms on the relationships between maternal urine parabens, including methyl, ethyl, and propyl parabens, and birth outcomes, including birth weight, gestational age, and birth height. Other studies have analyzed the effects of GSTM1 and GSTT1 on the associations between birth outcomes and the levels of other phenols, including mercury, perfluorinated compounds, and bisphenol A. However, this is the first study to analyze the effects of GSTM1/GSTT1 on the associations between birth outcomes and parabens levels.Our results showed the strong correlation between methyl and propyl parabens. This may be because they are the most two common parabens , and theIn the analysis of the association of combined GSTM1 and GSTT1 genotypes and urinary paraben concentrations with birth outcomes, we found that methyl and propyl paraben concentrations were positively associated with birth weight in both null genotypes. Moreover, this study’s results showed a significant interaction between GSTM1 and methylparaben for their outcomes of birth weight and propylparaben for their outcomes of birth weight, gestational age, and birth height. Furthermore, there were no significant interactions with any exposures and GSTT1.To date, several studies have investigated interactions between genetics and the environment. Chemical substances can increase the levels of polycyclic aromatic hydrocarbon DNA adducts by increasing the activity of enzymes such as GSTM1 and GSTT1 to hinder fetal and placental cellular regulation ,35. OxidSome studies have investigated the interactions of phenol exposure and birth outcomes with GSTM1 and GSTT1-null genotypes. Kwon et al.  reportedOur study is not without limitations. First, the number of subjects was smaller than those in other studies that investigated the relationships between chemicals, including PFCs, mercury, and bisphenol A, and birth outcomes in mothers with GSTM1 and GSTT1-null genotypes. Second, maternal urine was used as a proxy to determine fetal paraben exposure. Third, follow-up after birth has yet to be completed. Fourth, no comparisons with other studies could be made, as this is the first study to investigate individuals with GST null genotypes in this setting. Fifth, this study used a one-time urine sample, and measurement error should be considered in interpretation. Sixth, this study had no assessment of covariates related to socio-economic status. Lastly, we evaluated the single relationship between one exposure and birth outcome. One exposure can influence a physiological effect of other exposure. Therefore, our finding needs to be confirmed using a mixture analysis considering multiple correlated exposures.Regarding the strengths of this study, it is the first to explore the relationships between paraben levels and birth outcomes in mothers harboring GST polymorphisms. Moreover, this study analyzed each paraben  separately and adjusted for important confounders that may have affected paraben exposure or birth outcomes. Future studies are warranted in order to determine the details of the relationships between paraben concentrations and birth outcomes in individuals with GST polymorphisms.This study results showed a relationship between prenatal exposure to parabens and birth outcomes such as birth weight, gestational age, and birth height in pregnant women with GST polymorphisms. The maternal urinary concentrations of methyl and propyl parabens were positively associated with birth weight in mothers with the GSTM1 and GSTT1-null genotypes.


Women’s morphological femininity is perceived to develop under the influence of sex hormones and to serve as a cue of estradiol level, fertility and health in mating context. However, as the studies on direct relationship between femininity and sex steroid levels have reported mixed results, it is still not well understood what factors contribute to inter-women variation in morphological femininity. Epidemiological studies show that indicators of adverse conditions during intrauterine growth and development in utero, such as low birthweight or relative thinness at birth, influence women’s physiology ovarian functioning and may be associated with life-time exposure to estradiol in women. Thus, here we tested if birth parameters are also related with the level of morphological femininity in adult women.One hundred sixty-five healthy women of mean age 28.47 years (SD = 2.39) participated in the study. Facial femininity was estimated based on facial width-to-height ratio (fWHR) and facial shape sexual dimorphism measured in the photos. Body femininity was estimated based on waist-to-hip ratio (WHR) and breast size. Birth weight and birth length were obtained from medical records and ponderal index at birth was calculated. No relationship between birth parameters and facial or body femininity in women of reproductive age was found, also when controlled for adult sex steroid levels and BMI.The results suggest that, although previous research showed that birth parameters predict reproductive development and adult oestradiol level, they do not explain the variance in morphological femininity in women of reproductive age, trait that is thought to be a cue of a woman’s estradiol level and fertility in mating context. Sexual dimorphism, one of the most studied morphological traits as a putative cue to heritable fitness benefits , 65, 74,However, although inter-individual differences in the level of morphological femininity have been argued to be linked with variation in reproductive hormone levels , researcOne explanation for these equivocal results may be that morphological femininity is rather related with life-time exposure to estradiol levels, but not necessarily with adult estradiol level, that may be altered in response to changes in BMI , diet 1, 37, levThus, the aim of this study was to test if birth size contributes to the variation in women’s facial and body femininity in reproductive age. As intra-uterine development impacts adult physiology and may underlie life-time exposure to estradiol level, we hypothesized, that birth weight and adiposity  are positively related with facial and body femininity in women of reproductive age. As adult body adiposity  and sex r = 0.38, p < 0.001; 95%CI ) and with body WHR , but not with breast size . Similarly, facial SD was positively related with body WHR , but not with breast size  or E2 . We also found no relationship between birth ponderal index and tT  or E2 . None of the facial or body femininity measurements was related with testosterone or estradiol levels (in each case p > 0.05). Furthermore, BMI was not related with birth weight  or ponderal index  but was positively related with fWHR , facial SD , WHR , and breast size .The results of zero-order correlations between birth parameters and femininity measurements are presented in Table The results of multiple regression analyses showed no relationship between birth weight and facial or body femininity, also when controlled for pregnancy week at birth, BMI and sex steroids levels Table . The conSimilar results were obtained for the relationship between ponderal index and measures of facial and body femininity, also when controlled for pregnancy week at birth, BMI and sex steroids levels Table . The resp = 0.51), face sexual dimorphism (t (105) = − 0.30, p = 0.76), body WHR (t (105) = 0.22, p = 0.82) or breast size (t (105) = 0.43, p = 0.66). Also, comparison of women from the lowest and the highest tercile of birth ponderal index showed no difference in fWHR (t (91) = − 0.69, p = 0.49), face sexual dimorphism (t (108) = − 0.66, p = 0.51), body WHR (t (108) = − 0.23, p = 0.82) or breast size (t (108) = − 0.84, p = 0.40).Comparison of women from the lowest and the highest tercile of birth weight showed no difference in fWHR (t (105) = 0.67, The results of our study showed that birth parameters were not related with morphological femininity in women of reproductive age. The results were similar when controlled for pregnancy week at birth, adult BMI and sex steroids levels. Thus, although previous research showed, that birth parameters predict women’s reproductive physiology , 13, ovaThe results of our study are in line with the studies that failed to show the relationship between birth size and adult sex hormone levels , 57, 75.Furthermore, the results of this study are in line with findings suggesting no or weak relationship between facial sexual dimorphism and sex hormone levels or health , 65, queThe only variable related with facial and body femininity in our study was BMI. Similar relationship was shown in a number of studies on facial sexual dimorphism, showing that BMI is positively associated with fWHR in adults , 43, 46.Lack of the relationship between birth parameters and morphological femininity in women of reproductive age may also result from the fact that not all children with intrauterine growth retardation (IUGR) are characterised by low birth weight (LBW). Some children are smaller at birth due to their genetic predispositions, not unfavourable intrauterine environment, and these children may not be at higher risk of developmental disturbances resulting in lower health in adulthood. Similarly, some children with IUGR may not be characterised with low birth weight, due to their genetic predisposition for greater height . FurtherIt is also possible that some other factors, not controlled in this study, influencing a woman’s physiology during development , might impact the hypothesized relationship between birth parameters and morphological femininity. Such interaction among characteristics resulting from foetal programming (e.g. adult estradiol level) and from adult lifestyle, may be important in determining the relationship between size at birth and adult morphological femininity. For instance, in some countries, low birth weight is not related to an increased risk of atherosclerosis  or breasThe results of the study did not confirm our hypothesis on the positive correlation between birth parameters, indicating conditions during prenatal development that are related with life-time exposure to estradiol level, and facial and body femininity in women of reproductive age. As such, our study adds to a growing literature that questions current suggestions about the signal content of morphologically feminine traits.age = 28.51, SD = 2.37), recruited via social networks and information in the local media. Present study is one in a series, examining the relationship between women’s appearance, facial and body morphology and a wide range of measures of a woman’s biological condition. Data collected in the main project included information on hormone levels, oxidative stress and inflammation markers, blood biochemical markers, morphological measurements, and birth parameters. The protocol used to recruit participants and collect data was approved by the local ethical committee. For each participant, the general purpose of the study was explained and the written consent was obtained for participation in the study and use of data for scientific purposes.This research was a part of a broader project on women’s health, that included 209 participants , nulliparous, no current or chronic diseases, no hormonal contraception and no medication for at least 6 months before the study. Furthermore, only women born at term (37–41), with no evidence for a syndromic, chromosomal, or infectious aetiology of low birthweight were included. Participants were invited to participate in the study between the second and the fourth day of menstrual cycle (early follicular phase).3 and expressed in kg/m3. Ponderal index may be a better indicator of a newborn nutritional status than birth weight, in the same way that BMI (rather than adult body weight) better reflects adult nutritional status.Facial photographs were taken and body measurements were conducted. Women completed the study survey, containing questions on date of birth, past and current health problems, including hormonal disorders. Biochemical blood tests were also performed in order to verify general health status, and fasting blood sample was collected for hormonal analyses. Data on birth weight and length were obtained from the participants’ personal “health books”, containing records about birth size, health condition at birth, and any parturition problems. Ponderal index at birth was calculated as birth weight/(birth length)N = 5); 2) chronic health problems (N = 11); 3) being born as a twin (N = 2); 4) preterm births (N = 6); 5) incomplete data on birth parameters (N = 19); 6) testosterone level exceeding 3SD of the mean testosterone level (N = 1). The final sample included 165 women of mean age 28.47 years (SD = 2.39).Based on collected information and health measurements 44 participants were excluded from the analyses due to the following reasons: 1) other than early follicular phase of menstrual cycle  and facial shape sexual dimorphism  measured in the photos. Face images were taken under standardized photographic conditions with digital still camera (Nikon D7100 with Tamron SP AF 17-50 mm F/2.8 XR Di II LD IF camera lens). Camera-to-head distance and camera settings were held constant. Participants had no make-up, were asked to have a neutral facial expression, to remove glasses or earrings, and to wear a hairband.https://osf.io/98qf4/; R script for analysing sexual dimorphism scores following Scott et al. [age = 27.67 years, SD = 3.14 years) and 50 female  faces (recruited from the same population) were used to build the model used to calculate sexual dimorphism scores.fWHR was measured as the distance between the left and right boundary of the face (width) divided by the distance between the upper lip and the highest point of the eye-lid (height), following Stirrat & Perrett , using Wt et al.  and Komot et al. ). Lower Body femininity was estimated based on waist-to-hip ratio (WHR) and breast size. Circumferences of waist, hips, breast and under breast were measured to the nearest millimetre. WHR was assessed by the circumference of the waist divided by the circumference of the hips. WHR is inversely related with femininity. Breast size was assessed by the ratio of circumference of the breast to the under-breast circumference. Breast volume is positively related with femininity. Height and weight were measured and BMI was calculated. All measurements  were taken twice and means of the two measurements were used in the analyses.In order to evaluate serum testosterone and estradiol levels fasting blood samples were collected into EDTA Vacutainers (BD®). Serum was separated by centrifugation then portioned into micro-tubes and stored at -80 °C until analyses.Serum total testosterone (tT) concentration was assayed using ELISA commercial kits (Demeditec Diagnostics GmbH® cat. no. DE1559). Samples preparation and assay procedure were performed in accordance with the manual supplied with the kit. Samples, standards (included in each assay) were measured in duplicate. Optical density (absorbance) was measured using microplate reader  set to 450 nm. Testosterone concentration was calculated from the standard curve and expressed in ng/ml. Inter- and intra-assay coefficient of variability (CV) were less than 10% and less than 4.2% respectively, indicating good test precision and repeatability.Estradiol level (E2) was measured on Roche Cobas analyser in a certified analytical laboratory (DIAGNOSTYKA®), using Roche’s technology (ElectroChemiLuminescence methods). The serum estradiol concentration was expressed in pg/ml.The values of ponderal index, fWHR, WHR, breast size, BMI, E2 and tT levels were log-transformed for normalisation purposes, due to their strongly skewed or leptokurtic distribution.First, to explore the data we run zero-order correlation analyses between birth parameters and femininity measurements. Additionally, we ran zero-order correlation analyses between various measures of morphological femininity and between the measures of femininity and sex steroids levels. Subsequently, we verified if the relationship between birth parameters and femininity can be detected when controlled for the potential confounders that might impact femininity (body adiposity and adult sex hormones levels) or birth weight (pregnancy week at birth). Thus, we ran four multiple regression analyses with the measurements of femininity as the dependent variables and birth weight, pregnancy week at birth, E2, testosterone, and BMI as predictors. Another four regression analyses were run for the femininity measurements as dependent variables and birth ponderal index, pregnancy week at birth, E2, testosterone and BMI as predictors.3) and the third tercile (> 22.16 kg/m3) of birth ponderal index, in terms of morphological femininity, using t-test.In order to verify possible non-linear relationship between birth parameters and adult morphological femininity, we have also divided participants into three groups, based on terciles of birth weight, and compared the values of femininity measurements between women from the first (< 3200 g) and the third tercile (> 3555 g), using t-test. We have also conducted similar analysis for birth ponderal index, comparing women from the first (< 19.97 kg/mp < 0.05.Analyses were performed with Statistica 12.0 software. The results were interpreted as statistically significant if


The Nagasaki and Würzburg exchange and cooperation in the field of medicine goes back to the Würzburg physician and botanist Philip Franz von Siebold who lived in Nagasaki from 1823 to 1830 and again from 1859 to 1862. He was born into a famous Würzburg physicians’ family which provided Würzburg Medical Faculty with chairs in surgery and gynecology. Philip Franz von Siebold joined the Dutch East-Indian company in 1822 and arrived on the island of Dejima in the bay of Nagasaki on August 11, 1823. The Dutch settlement on Dejima was the only foreign settlement permitted in Japan until the mid nineteenth century. During his first stay, Philip Franz von Siebold was allowed to treat Japanese patients and to communicate with Japanese physicians such as Keisuke Ito, Mizutani Sugeroku, Ōkochi Zonshin and Katsuragawa Hoken on the Nagasaki mainland. He is thus considered by Nagasaki Medical Faculty as one of its founders together with the Dutch physician Johannes Lijdius Catharinus Pompe van Meerdervoort who lived in Nagasaki from 1857 to 1862 and established the first Western-style hospital and a formal medical school in 1861. Kusumoto-Ine, the daughter of Philip Franz von Siebold with his Japanese wife Sonogi O-Taki, attended this medical school and became a well-known Japanese physician while his German son Alexander Gustav von Siebold lived in Japan from 1870 to 1911 and became a cofounder of the Japanese Red Cross.Based on this history Nagasaki and Würzburg University started a formal student exchange program in 1996, signed by the President T. Berchem, the Dean K. Wilms, the President T. Yokoyama and the Dean S. Nagataki. Since then the academic exchange has been promoted particularly by Prof. C. Reiners, Prof. M. Tomonaga, Prof. S. Yamashita and the president Prof S. Kono. In 2006, a formal academic cooperation agreement has been signed by the President A. Haase and the President H. Saito. Over the years and partly funded by DFG and JSPS a number of bilateral research projects with reciprocal research stays in particular between the two psychiatric departments  and recently the two anatomy institutes (I. Takashi) have been successfully performed  established formal bilateral postdoc externship program funded by the two medical faculties. Both medical faculties and in particular the two psychiatric departments strongly believe that to overcome the challenges in a global world as highlighted by the recent SARS-CoV-2 pandemia, strategic partnerships including the educational, research and clinical level are essential.One major challenge during the current SARS-CoV-2 pandemia for all people and in particular for people with neuropsychiatric disorders is social isolation. Modern digital media have been employed to compensate for this at least to a certain degree and it is expected that they will change psychiatry for good (Gründahl et al. We therefore hope that the joint Nagasaki–Würzburg approach initiated by visionaries like Philip Franz von Siebold and his Japanese colleagues 200 years ago may serve as a blueprint for numerous other bi- and multilateral approaches to global challenges in the face of isolationistic and nationalistic approaches during the current SARS-CoV-2 pandemia.


Urachal cyst is an exceptionally rare disease in children caused by the incomplete obliteration of the urachal remnant. Urachal cysts seldom cause symptoms unless a secondary infection occurs. The symptoms of an infected urachal cyst are nonspecific and may be similar to acute appendicitis or other acute abdominal conditions. However, complications attributable to a delayed diagnosis can endanger the life of a patient.A 5-year-old boy presented with a 3-day history of severe intermittent lower abdominal pain.Infected urachal cyst.The patient was treated with surgical resection of the urachus, followed by intravenous antibiotics during the hospitalization.The patient was discharged without incident 7 days after the operation. With his follow-up in our out-patient department, he recovered well without any sequelae in the 6 months post-surgery.We suggested using the abdominal echo scan to differentiate the urachal cyst because of its high sensitivity and nonradioactive characteristic, and computed tomography is a typical diagnostic tool for urachal cysts. The mainstream management of an infected urachal cyst remains surgical excision. Complete excision of urachal cysts is relatively easy in a pediatric patient and the risk of subsequent infection is low; however, patients tend to have a low, although possible, risk of potential malignant transformation over their lifetimes. A urachal cyst is a congenital urachal anomaly attributable to the incomplete regression of the urachus during development in which both ends of the canal close. The incidence of urachal anomalies is thought to be rare. Asymptomatic urachal remnants may be present in up to 2% of the general population according to autopsy studies, but they are rarely symptomatic. Most urachal cysts are asymptomatic unless an infection occurs. In this paper, we report the case of a 9-year-old boy who received a diagnosis of an infected urachal cyst.The urachus, which connects to the urinary bladder, is a tubular embryonic structure derived from the allantois. Generally, the urachus is obliterated and forms a fibrous core before birth; however, some studies have suggested that this process may occur postnatally.Surgical management for urachal cysts used to be the standard of care. However, there has been more recommendation for observation and regular follow-up for urachal cyst in recent 1 decade. For an infected cyst in children, the necessity of surgical management is still the subject of debate. In this case, the patient received excision surgery and recovered smoothly.This case report was approved by the Institutional Review Board of Cathay General Hospital, Taipei, Taiwan. Informed consent was obtained from the patient and his parents.23) and C-reactive protein level (0.569 mg/dL). Other blood test results were in the normal range. Abdominal ultrasonography (US) revealed a 1.9 cm hypoechoic mass beneath the midline of the abdominal wall at the dome of the bladder . A plain abdominal X-ray revealed stool impaction, and a rectal enema was administered. The urine analysis results were normal. However, the symptoms persisted after he returned home. He was then brought to hospital B for further evaluation and care the next day . Urine analysis was performed again, results were normal, and probiotics were administered. Because of continuing abdominal pain, he was subsequently brought to our emergency room 2 days later . Examination revealed periumbilical tenderness and a fatigued general appearance, but the condition of the patient was otherwise normal. His body temperature was 36.4°C. Lab results indicated an elevated leukocyte count . In some studies, urachal cysts have been determined to be the most common urachal anomaly, even more frequent than patent urachus in children,–10 Urachal cysts were found in 1:5000 autopsies and occurred at a 3:1 male-to-female ratio. In another study, the ratio of boys to girls was 1.2:1 and of men to women was 2.0:1. Vincenzo et al, however, arrived at the opposite finding, concluding that a urachal cyst was the most common in girls .Urachal anomalies are related to the failure of the developing urachus to completely obliterate its lumen. Urachal anomalies can be diagnosed at any age, and their natural course and predisposing factors are still unknown. Although the actual incidence is unknown, urachal anomalies are thought to be rare. Urachal cyst is one of the congenital urachal anomalies attributable to the incomplete regression of the urachus during development where both ends of the canal close, most commonly at the lower third of the urachal tract. Urachal anomalies may close normally after birth but then reopen in association with pathologic conditions that are often categorized as acquired diseases.,13 Urachal cysts present a variety of clinical manifestations that include localized periumbilical or lower abdominal pain, fever, omphalitis, urinary symptoms, or a painful and palpable mass.,6,8,13 Abdominal pain is the most common symptom, according to most studies. The acute abdominal symptoms can be indistinguishable from those of acute appendicitis, Meckel diverticulitis, or peritoneal irritation after perforation of a viscus into the peritoneal cavity. However, a urachal cyst may sometimes occur as an abdominal mass or be noted unexpectedly. Umbilical discharge has sometimes been noted, most typically in infant patients.Most patients with urachal anomalies, except those with a patent urachus, are asymptomatic. Symptoms often accompany a secondary infection, which leads to early discovery. Urachal remnants are frequently diagnosed because ultrasound examination is often used as the screening tool. CT or US may reveal a fluid-filled cavity in the midline lower abdominal wall. Once an infection is present, the US image can reveal the complex echogenicity, and the CT image can reveal inhomogeneous attenuation using variable contrast enhancement in and around the disease process. Calcification has seldom been noted in urachal cysts but can be common in urachal malignancies. Some patients receive confirmed diagnoses of a urachal cyst after surgical exploration. A high risk of other related genitourinary anomalies may occur in relation to the urachal remnant. Retrograde urethrography or voiding cystourethrography are used for characterizing urinary bladder patency.To diagnose urachal cyst, a thorough medical history and physical examination are crucial. A definitive diagnosis often requires the use of computed tomography (CT). However, US is the first-line diagnostic tool and is thought to be effective especially for children. and is usually caused by the migration of bacteria, particularly Staphylococcus aureus, from the umbilicus. In our case, the pus culture taken from the infected urachal cyst contained E coli, which may have had a colorectal origin, and is commonly present in the urachal cyst fluid. Malignancy in adulthood is a concern. Malignancy in a pediatric patient is very rare. The first report of a mucinous urachal neoplasm in a urachal cyst in a pediatric patient was published in 2014. Since then, studies have reported a high rate of malignancies, contrary to the conventional view. Adenocarcinomas related to the urachal remnant remain uncommon, and most patients with simple and asymptomatic lesions do not appear to benefit from prophylactic excision.Complications of urachal cysts are often related to infection. If unrecognized, the infected cyst can perforate into the bladder or peritoneal cavity. This can cause peritonitis and the formation of an enteric fistula,,8,17,18 However, spontaneous resolution with nonsurgical management is likely with urachal cysts  in asymptomatic patients younger than 6 month-old. Observation remains a feasible management strategy for asymptomatic young patients because the accessibility of modern US enables a comprehensive follow-up strategy for the first 6 month to year of life. Early intervention (<6 month) should be reserved for patients with documented urine draining from the urachus or a documented abscess.Surgical management for urachal cysts used to be the standard of care because it could prevent secondary infection and potentially malignant changes later in life.,20 Since 2008, some studies have suggested nonsurgical management, even for infected urachal cysts. Initial drainage of the infected urachal cyst and/or combined antibiotic use with subsequent echography follow-up is sufficient. A study reported that an infected urachal cyst resolved spontaneously. Another study suggested that continuous observation with periodic ultrasound examinations was not necessary for asymptomatic cases because the urachal remnant disappears with age. Long-term follow-up into adulthood in a large patient cohort is necessary to effectively evaluate the necessity of removing the urachal cyst or a urachal remnant.The necessity of surgical management for children with infected urachal cysts is still the subject of debate. Intravenous antibiotic treatment is mandatory for an infected urachal cyst. A review of the literature indicates a 2-stage approach: drainage and antibiotics first and then excision.Most of the existing studies have discussed urachal remnants  other than urachal cysts separately. The clinical manifestation of urachal cysts is often nonspecific and diverse as in our case where the patient complained of acute abdominal symptoms but was misdiagnosed at the initial 2 clinical visits. It can be difficult to make a definite diagnosis through physical examination and medical history evaluation. Imaging methods, such as US and CT, are required. We should also be mindful of the differential diagnosis. In our review of the literature, the optimal management of an infected urachal cyst remained inconclusive. Because the spontaneous regression rate of urachal cysts in older children is not as high as in infants, carcinogenesis can be a risk, secondary infection is a possibility, and pediatric urachal excision can be completed with relative ease, we suggest that a complete urachal excision remains the best choice.Resources: Lung-Huang Lin, Shih-Pin Hung.Supervision: Lung-Huang Lin, Shih-Pin Hung.Writing – original draft: I Shan Tsai.Writing – review & editing: I Shan Tsai, Lung-Huang Lin, Shih-Pin Hung.


Background and objectives: The aims of this study were to examine the relationship between neurological outcomes at 3- and 6-months corrected age with the neurodevelopmental outcome at 3 years of age; to identify the perinatal/neonatal risk factors for poor neurodevelopmental outcomes at 3 years of age. Materials and methods: In our single-centre longitudinal cohort study, of the 73 consecutive infants admitted to our Neonatal Intensive Care Unit (NICU), 49 infants (80%) received both Hammersmith Infant Neurological Examination (HINE) at 3- and 6-months corrected age and Bayley–III neurodevelopmental assessment at 2–3 years chronological age. At 3 months follow up, 8.2% had suboptimal scores (below 10th percentile) on the HINE. At 6 months follow up, 4.1% had suboptimal scores (below 10th percentile) on the HINE. The means(±SD) for Bayley-III cognitive, language, and motor subscales were (96.3 ± 9.8), (99.9 ± 11.9), (93.2 ± 9.9). Results: At 3 months corrected age, higher total HINE scores and subscores for function of cranial nerves, posture, tone, were associated with better cognitive scores while poorer scores for function of cranial nerves, posture, movements, tone, and total HINE score were associated with lower motor scores. Infants with a HINE subscore of function of cranial nerves in the suboptimal range have three times higher odds of having a motor delay. Infants with a HINE subscore of function of cranial nerves in the suboptimal range have more than two times higher odds of having a language delay. At 6 months corrected age, poorer scores for function of cranial nerves, movements, tone, reflexes, and total HINE score were associated with worse Bayley-III motor scores whilst infants who have a total HINE score and a subscore of reflexes in the suboptimal range have four and seven times, respectively, higher odds of having a motor delay. Conclusions: Early identification of infants at risk for adverse long-term outcomes is essential in introducing early intervention therapies for optimizing neurodevelopmental outcomes. Over the past two decades, there has been extensive literature on babies with extremely low birth weight ELBW < 1000 g) and/or extremely preterm EP < 28 weeks), documenting improved survival mainly due to regionalization of perinatal care, improved technology, and better understanding of their pathophysiology and specific needs [000 g and8 weeks, Thus, comprehensive examination of ELBW/EP infants should include both neurological and neurodevelopmental assessments for fine-gross motor, cognitive, and language skills but also for social-emotional and adaptive behavior functioning, in order to provide the opportunity for enrollment to follow-up program, individualized family training, identification of infants eligible to early intervention services, and yield improved long-term outcomes ,10.Hammersmith Infant Neurological Examination (HINE) is a well-studied neurological exam in high-risk infants and can been used both for clinical and research needs; its sequential use also allows the identification of early signs of cerebral palsy and other neuromotor disorders .Among the most widespread tools for neurodevelopmental assessment, the Bayley Scales of Infant and Toddler Development III comprises of three distinct composite scores  allows a more accurate assessment of specific developmental domains, and contributes to an in-depth approach to premature infants’ follow-up . HoweverAlthough, HINE score and individual items are predictive of motor outcomes, it is unclear whether those scores and items are associated with specific deficits in cognitive, fine-gross motor, or expressive-receptive language domains of neurodevelopment.Therefore, the aims of this prospective study in a cohort of preterm ELBW infants were firstly, to examine the relationship between neurological assessments at 3 and 6 months corrected age with neurodevelopmental outcome at 3 years of age; secondly, to identify the perinatal/neonatal risk factors for poor neurodevelopmental outcomes at 3 years of age.We implemented a single-center longitudinal cohort study. The study design was approved by the Hospital’s Ethics Committee  and written informed consent provided by parents for their infant to participate in the study.Inclusion criteria included: birth weight ≤1000 g (ELBW) and/or gestational age ≤ 29 + 6 weeks. Infants with genetic abnormalities and/or neurosensory disabilities  were excluded from the study.All the 73 ELBW infants consecutively admitted to our NICU from January 2014 to December 2014, 62 (84%) were discharged home alive. At baseline, 39 infants were males (53.4%) and 34 females (46.5%). Of all 73 infants recruited, 61 infants entered the study, 1 was excluded due to genetic abnormalities and 11 died, 56 (91.8%) returned for the 3 months corrected age follow-up visit, 54 (88.5%) for the 6 months corrected age follow-up visit. In total, 49 infants (80%) received both HINE neurological assessments (at 3- and 6-months corrected age) and Bayley–III neurodevelopmental assessment at 2–3 years chronological age .Study data were collected prospectively and included: infant’s sex, gestational age (weeks), and weight (grams) at birth, head circumference. Other neonatal variables included: Apgar scores at 5 min, singleton, twin or multiple pregnancy, mode of delivery, being small for gestational age (SGA), and length of hospital stay. Clinical variables such as respiratory distress syndrome, duration of invasive mechanical ventilation, days on oxygen, postnatal surfactant treatment, sepsis, patent ductus arteriosus (PDA), bronchopulmonary dysplasia, necrotizing enterocolitis, retinopathy of prematurity (ROP), and complications during the hospital stay such as intraventricular hemorrhage the highest grade documented (IVH), periventricular leukomalacia (PVL) were recorded. Maternal characteristics included chorioamnionitis, antenatal corticosteroid therapy, preeclampsia/eclampsia, chorioamnionitis, gestational diabetes, and intrauterine growth retardation (IUGR).Gestational age was determined from the date of the last menstrual period and was confirmed by early ultrasound examination. BW was measured with an electronic scale. Infants were classified as small for gestational age according to the Fenton Growth Chart . NeonataInfants were scheduled to be examined at 3, 6 months corrected age and 36 months chronological age. Hammersmith Infant Neurological Examination (HINE) was performed at 3- and 6-months appointments, after a standardized physical examination. An experienced neonatologist and a trained pediatric physiotherapist performed neurological and neurodevelopmental assessments.HINE assessment consists of 26 items exploring the areas of cranial nerve function, posture, quality, and quantity of movements, tone, and reflexes/reactions. Each item is scored individually; the examiner marks the most relevant of the four columns scoring 0,1,2,3, and scores can be added to achieve a global optimality score, range: 0–78. The assessment has been standardized in very preterm infants between 6- and 15-months corrected age , and alsAt 36 months chronological age a pediatric physiotherapist, unaware of the child’s perinatal history, applied neurodevelopmental assessment using Bayley-III scales to all infants. The third edition of the Bayley Scales of Infant and Toddler Development is a valid, reliable, and standardized assessment tool of developmental delay . It encoBayley III, in all three composite index scores, has index mean scores of 100(SD ± 15). An index composite score of <70 (>2 SD below the mean) is defined to indicate severe impairment, while an index composite score of 71–85 (>1 SD below the mean) is defined to indicate mild impairment. According to Bayley III, index composite scores ≥86 indicate normal development .2 or Fisher exact tests. To determine the relation between continuous variables, the Mann-Whitney or the t test was used depending on the distribution. Fisher exact test was used to examine the association between HINE suboptimal subscores and total scores at 3- and 6-months corrected age and developmental delay on Bayley III at 3 years chronological age. Linear regression analysis was used to examine the relation between perinatal/neonatal characteristics of ELBW infants and Bayley III scores in motor, cognitive, and language domains at 2-3 years chronological age. Results are reported as relative risks (RR) and 95% confidence intervals. p < 0.05 was considered statistically significant.The Statistical Package for the Social Sciences  was used for data analysis. The relation between categorical variables was investigated using χThe mean age at testing was at a corrected age of mean(±SD) 3.1(±0.85) months and 6.8(±1.44) in the first and second follow up visit with HINE examination, respectively, while the mean age at Bayley–III testing was at 30.9(±7.29) months chronological age.As shown in Four infants had cerebral palsy, and according to Gross Motor Function Classification System (GMFCS) one was diagnosed with hemiplegia/level I, 2 had diplegia/level I, and one had quadriplegia/level III .Neurodevelopmental outcomes for cognitive, language, and motor domains are reported in Overall at 3 months, follow up 8.2% had suboptimal scores (below 10th percentile) on the HINE with 4.1% for Function of cranial nerves, 2% for Posture, 4.1% for Movements, 10.2% for Tone, and 2% for Reflexes and reactions.At 6 months follow up, 4.1% had suboptimal scores (below 10th percentile) on the HINE with 12.2% suboptimal for Function of cranial nerves, 6.1% for Posture, 4.1% for Movements, 8.2% for Tone, and 6.1% for Reflexes and reactions. The median (range) for the HINE at 3- and 6-months corrected age is reported on p = 0.082), posture , tone , and total HINE score , were associated with improved cognitive composite score, while poor total HINE score and sub-scores were associated with lower cognitive performance  scores for function of cranial nerves  were associated with increased risk of language delay , posture , movements , tone , and total HINE score  were associated with lower composite motor scores at 3 years of age , movements , tone , reflexes , and total HINE score  were associated with lower (worse) motor performance at 3 years of age  were associated with increased odds of motor delay on Bayley-III at 3 years of age , and HINE total score  were associated with increased risk of motor delay on Bayley-III , prenatal steroids (p = 0.002), PDA (p = 0.013), and necrotizing enterocolitis (p = 0.028). Additionally, necrotizing enterocolitis was found to influence both cognitive and language composite scores .Regression analysis indicated that significant independent factors at We studied the association of the HINE neurological examination at 3- and 6-months corrected age with Bayley-III in a cohort of EP/ELBW infants at two to three years of chronological age. At 3 months corrected age, higher (better) total HINE scores and higher HINE subscores for function of cranial nerves, posture, tone, were associated with better cognitive performance, while poorer scores for function of cranial nerves, posture, movements, tone, and total HINE score were associated with lower (worse) motor scores. Infants who have a HINE subscore of function of cranial nerves in the suboptimal range have three times higher odds of having a motor delay. The relationship between total HINE score/subscores and language scores is less clear but infants who have a HINE subscore of function of cranial nerves in the suboptimal range have more than two times higher odds of having a language delay.At 6 months corrected age, poorer scores for function of cranial nerves, movements, tone, reflexes, and total HINE score were associated with lower (worse) Bayley III motor scores whilst infants who have a total HINE score and a subscore of reflexes in the suboptimal range have four and more than seven times, respectively, higher odds of having a motor delay.According to the scientific literature, HINE is the most frequently used method internationally ; it is tOur study has demonstrated that both total score of the HINE and subscale scores alone were strongly related to motor outcomes in Bayley III test and this is consistent with previous findings using HINE, which has shown to be reliable in predicting motor outcome at 2 years of age in very preterm infants assessed between 9 and 18 months of age [We obtained reference cut-off values for the optimality scores at 3- and 6-months corrected age and found that ELBW infants with suboptimal HINE scores/sub-scores, as early as 3 months corrected age have higher odds for delay in the language domain. Accordingly, in a cohort of very preterm infants by Romeo et al. (2009), a high correlation, as early as 3 months onwards, between the HINE and the receptive and expressive language development at 24 months was reported .It has been reported that auditory processing abilities are well developed in infancy, and thus, infants with such deficits are at a higher risk for developing a language disorder . In the We found that poorer HINE subscores for function of cranial nerves, posture, tone, and total HINE scores were associated with worse cognitive scores. Posture subscale of the HINE tool assesses head, trunk arms, hands, legs, and feet posture and asymmetries in those parts of the body. According to de Groot, in early infancy, the development of adequate motor behavior and sensorimotor interaction could be disrupted by inadequate postural behavior, resulting in a faulty perception–action cycle, thus influencing later cognitive development . FurtherRecognizing early developmental delays is crucial for providing early intervention services because the brain is considered to be especially plastic in the phase occurring after the completion of neuronal migration. The processes of dendritic outgrowth and synapse formation are highly active during this phase . The majThe strengths of our study include the use of a longitudinal prospective cohort, along with very good follow-up rates.Our study did not involve a population-based cohort, which may limit the generalizability of findings to other populations. It did reflect care at a regional neonatal intensive care unit. Another limitation of the current study is its monocentric nature. A multi-center study could enable the recruitment of higher number of infants and subsequently stronger associations. It would also be beneficial in a longer-term follow up, as the predictive validity of Bayley-III is moderate for school-age outcomes ,33.The current study has considerable implications both in clinical and research settings. HINE at 3- and 6-months corrected age is associated with cognitive, language, and motor performance at 2–3 years in ELBW infants. HINE assessment is a simple, no-cost tool detecting delayed high-risk infants in motor, cognitive, and language domains. Early identification of ELBW infants at risk for adverse long-term outcomes is crucial for intervening early in life with physiotherapy/occupational therapy programs, optimizing neurodevelopmental outcomes and function. Furthermore, a systematic review of the literature targeted on the relationship between ELBW infants and alterations /deficits in their neuropsychomotor development would be recommended.


Low birth weight is defined as when a newborn weighs less than 2,500 grams within an hour of birth. Globally, it has been known that around 15.5% of newborns were below the normal level of weight at their birth and 95% of these infants lived in developing countries. The main objective of this study was to assess the prevalence and associated factors of low birth weight among newborns delivered at Butajira General Hospital, Southwest Ethiopia.  An institutional-based cross-sectional study design was employed. All 196 paired study participants (newborn-mother) who were born on a one-month duration of the data collection period were included in the study. A pretested questionnaire was used to gather pertinent information about mother and newborn along with measuring newborn birth weight.  Majority of mothers 175 (92.1%) were aged between 20 and 34 years, and 186 (97.9%) were married. About 169 (88.9%) were protestant religion followers. This study showed that the magnitude of low birth weight among study participants was 12.5%, and factors such as maternal medical complication during pregnancy, maternal MUAC less than 23 cm, and birth interval less than 24 months were significantly associated with low birth.  The study finding indicated that a significant number of newborns measured underweight which is below the normal level of weight at birth. The study identified factors such as maternal medical condition during pregnancy, maternal MUAC less than 23 cm, and birth interval less than 24 months. Based on study findings, we recommend health care officials, policymakers, key persons in the family, and volunteers to work on nutritional values particularly during pregnancy and before pregnancy. Spacing of birth is crucial to have healthy baby and healthy family even healthy society at large, so attention should be paid on family planning utilization. Low birth weight is defined by the World Health Organization (WHO) as weight at birth of less than 2,500 grams. Birth weight is an important indicator of health status of an infant and is a principal factor that determines the infant survival and physical and mental growth. It also indicates the past and present health status of mother , 2.LBW is considered as a single most important predictor of infant mortality, especially of death within the first month of life . It is aGlobally, neonatal mortality is 20 times more likely for LBW newborns when compared to normal weight newborns  and it increases sharply as birth weight (BW) decreases . NewbornLow birth weight a major public health issue in developing countries like Ethiopia. LBW leads to impaired growth of the infant resulting in a higher mortality rate and increased morbidity. LBW is also an important indicator of maternal and child health . A low bThis study will help to fill the information gap regarding prevalence and associated factors of LBW in Butajira General Hospital.This study may increase the awareness of health professionals such as midwives, nurses, gynecologists, and others about the major risk factors for LBW and help them to give a proper and evidence-based health service  in order to anticipate the problems associated with low birth weight.Furthermore, this study will provide valuable information for researchers for further study.To assess the prevalence and associated factor of low birth weight among newborns at Butajira General Hospital, South Ethiopia.To determine the prevalence of low birth weight among newborns at Butajira General HospitalTo identify factors associated with LBW among newborns at Butajira General HospitalThis study was conducted in Butajira General Hospital found in Gurage Zone, South Ethiopia. Butajira is a town and separate woreda in south-central Ethiopia. Located at the base of the Zebidar mass in the Gurage Zone, Southern Nations, Nationalities and Peoples' Region. The Town has an elevation of 2,131 meters above sea level. It is surrounded by Meskane woreda.Butajira General Hospital is situated 130 km south of the capital of Ethiopia, Addis Ababa, and 50 km to the west of Ziway town in the Rift Valley.The study was conducted from May 1, 2019, to June 1, 2019.The institution-based cross-sectional study design was conducted from May 1, 2019, to June 1, 2019.All mother-newborn pairs in Butajira General Hospital during the data collection period were the source population.The study population included selected mother-newborn pairs.All mothers who delivered live newborns during the study period were included in the study.The sample size was calculated by using a single population proportion formula by taking the prevalence of LBW which was 13.4%  and margn = (Z a/2)2p (1 − p)/d2, where n is the desired sample size, d is the desired precision 5% = (0.05), z is the standard normal distribution value at confidence level 95% = 1.96, p is the prevalence rate of low birth weight = 13.4%, (Zap/2)2 = (1.96)2 = 3.8416, P = 13.4% = 0.134, and d = (0.05)2 = 0.0025.A total sample size was determined as follows: n = 3.8416∗0.134∗0.866/0.0025 = 178.32.So, By taking nonresponse rate, 10% of 178.32 = 17.832.The total sample size was 178.32 + 17.832 = 196.15 ~ 196.A simple random sampling technique was used to select study participants.The dependent variable is low birth weight.Sociodemographic characteristics: maternal age, family monthly income, educational level, maternal occupation, religion, marital status, ethnicity, and residenceMaternal medical and obstetric characteristics: gravidity, parity, history of abortion, desirability of pregnancy, history of previous preterm birth, birth interval between current and previous last pregnancy, HIV status of the mother, ANC follow-up, number of ANC visits, and medical complication during pregnancy like malaria, PIH, and anemiaNutritional status of mother: average daily food consumption, additional diet during pregnancy, iron and folic acid supplementation during pregnancy, nutritional counseling during pregnancy, height of the mother, and MUAC of the motherGeneral maternal behaviors: history of alcohol drinking, cigarette smoking, and chewing chatNewborn characteristics: sex of the newborn, gestational age at birth, and birth orderLow birth weight is a weight of less than 2,500 g  irrespective of the gestational age .Gestational age is the duration of time measured from the first day of conception and expressed in completed weeks.Preterm birth is the birth of newborn before 37 completed weeks of gestational age.Gravidity is the number of pregnancy.Parity is the number of births or the number of children either alive or died.Birth order is the sequence at which birth occurs.Maternal behavior is a habit of smoking cigarette, drinking alcohol, and chewing chat.Birth interval is a time period between the current and previous last pregnancy.Data was collected by using a pretested, semistructured questionnaire with an interview type of data collection method, reviewing records from medical registration cards and actual measurement of some variables. Weight of newborn was measured immediately after delivery by using the neonate weight measurement scale. The questions were prepared in English and translated into Amharic language. The interview was in Amharic language, commonly used local language.The quality of data was ensured during collection, coding, entry into the computer, and analysis. Orientation was given to data collectors and supervisors. Nurses who are currently working are involved in data collections. Each mother was asked, and the cards also were checked for consistency, provision of full information, and appropriate documentation. Neonate birth weight was measured using a birth weight measurement scale which is usually used in a hospital; its measuring accuracy was checked by supervisors before actual measurement takes place. The questionnaire was pretested on a 5% sample size before data collection using women postnatal room. The filled questionnaire was cross-checked by team members for completeness and closely supervised by supervisors daily.Data were entered by using epi-data version 3.1, and SPSS version 24 was used for analysis. Logistic regression was used to see the association between an outcome variable and independent predictors.All 196 mothers with their paired newborns responded which gives a response rate of 100%. Majority of mothers 175 (92.1%) were aged between 20 and 34 years, and 186 (97.9%) were married. About 169 (88.9%) were protestant religion followers. Nearly above half of 107 (56.3%) had a monthly income of 3,000 Ethiopian birr, 124 (65.3%) lived in a rural area, 86 (45.3%) were housewives, and 67 (35.3%) were educated grade 1-8 (primary school).Most of the mothers 184 (96.8%) responded that their pregnancy was planned, and 180 (94.7%) were tested for HIV status and declared as negative. About 136 (71.6%) were gravid 2 to 5, 141 (74.2%) were Para 2 to 5, 80 (40.5%) had birth interval less than 24 months, and about 29 (19.0%) and 10 (6.5%) had a history of abortion and preterm birth, respectively. About 184 (96.8%) mothers had ANC follow-up, and from these, only 49 (26.5%) mothers had ANC visit greater or equal to four visits, 54 (29.4%) mothers had no iron and folic acid supplementation, and 46 (25.0%) mothers had no nutritional counseling during current pregnancy. About 129 (67.9%) mothers did not consume additional diet than usual during current pregnancy, and about 120 (61%) and 100 (51%) mothers had a height greater than or equal to 155 cm and midupper arm circumference less than or equal to 23 cm, respectively. In this study, about 50 (25%) mothers had a history of medical conditions, and from these, 14 (28%) mothers had hypertension, 12 (24%) had Anemia, 15 (30%) had malaria, and 11 (22%) mothers had other medical conditions. All mothers 196 (100%) participated in this study had no habit of cigarette smoking, alcohol drinking, and chewing chat.Of the total 196 newborns, 100 (51.6%) were female. Most of them 185 (94.4%) were born at a gestational age greater than or equal to 37 completed weeks. About 123 (64.6%) newborns were 2 to 5 birth order, and about 23 (12.5%) newborns were born with weight less than 2,500 grams.p < 0.0001 . This study finding is consistent with a community-based study done in Karaka, India . MothersAnother factor identified in this study was mother MUAC of a mother. Mothers whose midupper arm circumference measurement is 23 cm and above were at a lower risk of giving LBW newborns  than mothers who had midupper arm circumference less than 23 cm. This finding was in consistent with the study done in Addis Ababa .Since this study used a cross-section study design which could not show a cause-effect relation between variables, measuring newborn weight may induce bias of accuracy. The study used a small sample size due to the availability issues of participants.The study indicated that the magnitude of low birth weight among participants was high. Factors associated with LBW were maternal medical conditions, maternal midupper arm circumference, and birth interval between previous last pregnancy and the current one. Based on study findings, we recommend health sector officials, professionals, and government and nongovernment organizations to work on maternal nutritional status which directly affects newborn weight. In addition, keeping recommended birth spacing is required to advance both mother and infant.


To evaluate gestational age as a predictor of subsequent preterm birth.This was a retrospective birth cohort study to evaluate gestational age as a predictor of subsequent preterm birth. Participants were mothers who gave birth to their first two children in Western Australia, 1980–2015 . For each week of final gestational age of the first birth, we calculated relative risks (RR) and absolute risks (AR) of subsequent preterm birth defined as final gestational age before 28, 32, 34 and <37 weeks. Risks were unadjusted to preserve risk factor profiles at each week of gestation.The relative risks of second birth before 28, 32, and 34 weeks’ gestation were all approximately twenty times higher for mothers whose first birth had a gestational age of 22 to 30 weeks compared to those whose first birth was at 40 weeks’ gestation. The absolute risks of second birth before 28, 32, and 34 weeks’ gestation for these mothers had upper confidence limits that were all less than 16.74%. The absolute risk of second birth before 37 weeks was highest at 32.11%  for mothers whose first birth was 22 to 30 weeks’ gestation. For all gestational ages of the first child, the lowest quartile and median gestational age of the second birth were at least 36 weeks and at least 38 weeks, respectively. Sensitivity and positive predictive values were all below 35%.Relative risks of early subsequent birth increased markedly with decreasing gestational age of the first birth. However, absolute risks of clinically significant preterm birth , sensitivity and positive predictive values remained low. Early gestational age is a strong risk factor but a poor predictor of subsequent preterm birth. It is widely accepted that previous preterm birth is a strong risk factor for preterm birth in later pregnancies –4. EstimBecause previous preterm birth is a strong independent risk factor for subsequent preterm birth it can inform clinical management to prevent recurrence, such as screening for tocolytic treatment . Family Population-representative summaries of gestational age based on the gestational age in previous pregnancy requires follow-up of enough mothers over a period sufficient to observe successive pregnancies with an adequate number of births at early gestations. The availability of linked individual-level perinatal data and population-wide coverage of births provides an opportunity to produce such summaries. The aim of this study was to estimate the absolute risk of preterm birth by gestational age of the previous pregnancy, and to estimate the predictive ability of the gestational age of the previous pregnancy.This was a retrospective cohort study using probabilistically linked midwives notification data to estimate the absolute risk of preterm birth of the second birth based on the gestational length of the previous birth, for all notified singleton births in the state of Western Australia, 1980–2015.Participants were women who gave birth to their first two singleton children in the study region, during the study period. From a starting population of 964,015 births, we sequentially excluded 26,730 multiple births; 5,187 births with missing final gestational age; 121 births with missing parity; 712 births with duplicate parity; and 1,872 births with gestational ages < 22 weeks or > 44 weeks to minimise influence of undetected late induced abortions and gestational age recording errors at the extremes of gestational age . BecauseUnit records from the Midwives Notification System were obtained from the Western Australian Department of Health . The MidEthical approval for this study was obtained from the Human Research Ethics Committee, Department of Health, Western Australia (#2016/51).As the aim of our study was to evaluate gestational age of the first child as a predictor for preterm birth of the second child we reported “absolute risk” of preterm birth and standard prediction metrics. These metrics provide information on the absolute degree to which the gestational age of the first child indicates preterm birth of the second child. To evaluate gestational age of the first child as risk factor for preterm birth, we also report “relative risks”. Relative risk provides information on the degree to which the gestational age of the first child indicates preterm birth of the second child compared to an alternative scenario (reference category). Relative risk in this study is the multiple by which risk of preterm birth of the second child is greater given a particular gestational age of the first child, relative to risk if the first child was born at full-term .extremely preterm, very preterm, moderately preterm and preterm)  to 13-85%”. These are the results that we presented in the “Absolute risk of preterm birth of the second child by gestational age of the first child” sub-section of the Results in the original submission:“The absolute risk of second child birth before 32 weeks’ gestation ranged from 9.19% to 11.64% if the gestational age of the first child was between 22 and 30 weeks, and less than 7.13% if the gestational age of the first child was more than 30 weeks. The interval estimates of absolute risk of second child birth before 34 weeks’ gestation ranged from 13.85% to 16.74% if the gestational age of the first child was between 22 and 30 weeks, and was less than 12.32% if the gestational age of the first child was more than 30 weeks. In contrast, the absolute risk of second child birth before 37 weeks’ gestation was elevated for all gestational ages of the first child less than 37 weeks, ranging from 10.90% to 34.56%.”Again, we agree with the reviewer. The “increase” is notable. This “increase” is reported in our study via the “relative risk”, which we have interpreted in the “Relative risk of preterm birth of the second child by gestational age of the first child” sub-section of the Results. The reviewer is interpreting the change in absolute risk estimates, which by definition is a relative risk. In contrast, the results presented in the “Absolute risk of preterm birth of the second child by gestational age of the first child” sub-section of the Results should be interpreted in terms of their absolute magnitude. For every comparison investigated, predictions of second child preterm birth based on first child preterm birth were all < 35%, which is much lower than the probability from the flip of a coin. Of course, other metrics such as sensitivity and positive predictive value, are relevant but these were also very low (see the paragraph in our response to R1.3). R1.5. What was the risk of a second child being born preterm if the first child was born after 37% weeks, I cannot find this analysis and this is what all the numbers should be compared to and not to the preterm birth risk of a first born.Response to R1.5. We agree with the reviewer’s recommendation of the reference category, and stress that we did this in the original submission. The results are reported in Table 2. The relative risk estimates use full term (40 weeks) as the reference category. We have reported relative risks of preterm birth for the second child when the gestational age of the first birth was 22-30 weeks, 31-32 weeks, each week from 33 to 41 weeks , and 42-44 weeks. We are unsure if the reviewer is aware how to interpret “relative risks” and have provided the following example for their benefit . In this example we interpret the relative risk of 8.98, reported in Table 2. This relative risk of 8.98 is when the first child was born at 33 weeks of gestation and the second child is born <37 weeks. The interpretation is that the probability of second child birth <37 weeks when the first child is born at 33 weeks is 8.98 times the probability of second child birth <37 weeks when the first child is born at 40 weeks. R1.6. Was the interval between first and second pregnancies accounted for?Response to R1.6. This is a reasonable question given the dominance of etiological studies in the published literature. We agree with the reviewer that adjustment for potential confounders, including but not limited to interpregnancy intervals, are likely to attenuate the relative risks. However, our study was not an etiological study. The aim of our study was to evaluate gestational age of the first child as a predictor for preterm birth of the second child . Adjustment is not relevant and the rationale for this is described in the manuscript . We reported relative risks alongside absolute risks and prediction metrics for completeness. R1.7. In the section, what this study adds, the statement that the median age of the second child is at least 38 week including births for mothers when their first child was extremely preterm is misleading and again downplaying the increased risks. Of course the median age in 38 weeks if even 35% of the babies were born preterm. What was the median age of the babies born preterm after the first birth was preterm? Did that increase or decrease?Response to R1.7. We have removed this section to conform to the journal’s requirements. We made a small amendment to make the title less subjective. We did not intend for the statement to which the reviewer refers to be misleading and do not believe our results are misleading because the median gestational ages of the second birth by gestational age of the first birth were reported in the original and revised submissions for all gestational ages of the first birth (Table 2). We also did not limit our results to the median, and believe we were thorough and highly transparent in reporting multiple quantiles . To answer the reviewer’s specific question, the median gestational age of the second birth was 38 - 40 weeks and therefore changed very little. However, as we have previously stated (see response to R1.4) the “increase” or “decrease” in gestational age and chance of preterm birth across the range of gestational ages of the first birth is a relative measure of risk. Again, we agree with the reviewer that relative risks of preterm birth increase with decreasing gestational length of the first child. This has been reported by others as well. However, our results show that gestational length of the first child is a poor predictor of preterm birth of the second child (see Absolute Risks in Table 2 and prediction metrics in Table 3) despite it being a strong risk factor (see Relative Risks in Table 2). Response to Reviewer 2 (R2)We thank Reviewer 2 for the time and effort to review our manuscript. We have responded to each comment below. Where changes have been made to the manuscript we refer to line numbers corresponding to the Tracked Changes version of the resubmission. R2.1 General impression: This is the first large study assessing the confidence intervals of both the relative risk and the absolute risk of recurrent preterm birth after a first preterm singleton delivery. Strengths of the study include large population-based study with best obstetrical estimate of gestational age. Methods are well explained, data are clear, and the manuscript is easy to follow. Limitations include lack of analysis of other factors for subsequent preterm pregnancy beyond gestational age, long duration of data collection and lack of testing summary statistics in another population.Response to R2.1. We appreciate the reviewer’s positive comments and would like to raise attention to another strength of the study, namely that preterm birth has been long regarded as a risk factor for subsequent preterm birth but until now has never been evaluated as a predictor. We have addressed the limitations raised by the reviewer in the specific comments below. R2.2. The authors should state other factors that may affect the risk of recurrent prematurity and list relevant literature. Previous authors  have shown that other factors may affect the risk of recurrent preterm delivery, e.g., preeclampsia, maternal age <18 years, preterm labor vs. preterm premature rupture of membranes. If this information is not available, it should be listed as a limitation.Response to R2.2. We agree with the reviewer that there are multiple risk factors for “recurrent prematurity”, including those mentioned as well as genetic predisposition. We intentionally did not adjust for risk factors. We intentionally allowed the relative risks to reflect both direct effects of first child gestational age as well as indirect effects via other associated risk factors such as those mentioned by reviewer. If we included other risk factors, such as those mentioned by the reviewer, the model may improve its predictive performance but the predictive performance of first child gestational age (which is the primary subject of this study) would decrease. Our relative risk estimates therefore reflect the “total effect” of first child gestational age on risk of preterm birth of the second child. The aim of our study was to evaluate gestational age of the first child as a predictor for preterm birth of the second child  and therefore adjusting out the effects of other risk factors is not relevant .wa.gov.au/articleR2.3. The study lasted 35 years. It is likely that practice of pregnancy dating and prevention of preterm birth has changed during this time. Concerns were raised about validity of gestational age in the 1986 report listed on ww2.health.https://ww2.health.wa.gov.au/-/media/Files/Corporate/general-documents/Data-collection/PDF/Midwives_Validation_Study_1986.pdf). Specifically, the 1986 validation study states that the estimated gestation on the midwives notification was the same as that stated in the medical record for 79.4% of cases. The 1986 validation study did not report a measure of central tendency such as the mean absolute difference between the two dates or any useful measure of the degree of the difference. The most recent validation study was conducted in 2007 . The 2007 validation study reported that estimated gestation was correct for 468 of the 525 records (89%). More importantly, the 2007 study indicated that of the 57 records for which estimated gestation was incorrect, 50 of them were due to a rounding error whereby the midwives notification was rounded to the nearest whole week of gestation instead of rounding down to the completed week of gestation. If we are willing to accept such a small error, this would mean that only 98.7% of the midwives notifications had the correct gestational age.Response to R2.3. We acknowledge the reviewer’s concern and broadly agree with the reviewer. The long duration of data collection is both a strength and a limitation. The long duration allows us to observe larger numbers of births at early gestations (closer to viability) than if the duration of the study was shorter. However, a longer study period is accompanied by temporal changes. The validity of gestational age was mentioned in the 1986 validation study , with the latter more common earlier in the cohort period. The 1986 validation study states that birth dates were accurate for 99.7% of the records in the validation study  and last menstrual period dates were correctly transcribed for 97.1% of the records (on p19). The transcription of these data items is not a limitation. The main limitation is due to the loss of accuracy when using gestational age estimated from LMP compared to gestational ultrasound dating. The consequence of this limitation is likely to have resulted in wider confidence intervals (loss of precision), which we have now described as a limitation in the Discussion. We have also described secular changes in antenatal care as a limitation because this could have resulted in prevention of preterm birth. However, we note that the change in antenatal care between each mothers first birth and second birth is likely to be minimal because this interval is of the order of years not decades. Similarly, we mentioned the potential for changes in behavioural risk factors during the study period. Again, we anticipate that these limitations are likely to have resulted in wider confidence intervals (loss of precision), but this would not have changed our conclusion because estimates in Table 3 were already precise for sensitivity and positive predictive values.The additional description on limitations is now described in the Limitations paragraph, at lines 241 - 246 of the revised manuscript, which is the last paragraph of the Discussion section.R2.4. Lack of testing sensitivity, specificity and positive predictive value in another population is a major limitation. Values provided only apply to the population from which these data were calculated; they may not apply to a prospective current population.Response to R2.4. We thank the reviewer for raising this point. In response, we have now mentioned the fact that our results need to be replicated/tested in an independent study population at lines 246 - 250 in the Limitations paragraph, which is the last paragraph of the Discussion section. We have also referred to this limitation as the main limitation of our study. Our starting study population included almost one million births over more than three decades for the whole state. We do agree that this is a limitation but do not believe it is a major limitation because validation at this scale – replication of a study of this size in an independent population (which is needed to observe enough births at early gestations) - would be a significant undertaking, undertaken in a new study and separately published. R2.5. The discussion lacks a paragraph on limitations.Response to R2.5. Limitations were mentioned in the second last paragraph of the original submission, although we acknowledge that it was sparse. We have included further description of limitations at lines 241 - 250 in the Limitations paragraph, which is the last paragraph of the Discussion section in the revised version of the manuscript. R2.6. In the abstract, results, the second sentence is not accurate. It would be more exact to state the following: “... had upper confidence interval limit equal or less than 16.74%”.Response to R2.6. We appreciate the suggestion and have made the change recommended by the reviewer at line 31 of the Results section.R2.7. What is known on this subject: Several studies have reported the absolute risk of recurrence of prematurity after a first preterm delivery. It would be correct to state that the confidence intervals of the absolute risk of recurrence of prematurity after a first pregnancy have not been reported.Response to R2.7. We have removed this section to conform to the journal’s requirements but agree with the reviewer’s comment nonetheless.AttachmentResponse to Reviewers.docxSubmitted filename: Click here for additional data file. 11 Jan 2021Re-evaluation of gestational age as a predictor for subsequent preterm birthPONE-D-20-29280R1Dear Dr. Pereira,We’re pleased to inform you that your manuscript has been judged scientifically suitable for publication and will be formally accepted for publication once it meets all outstanding technical requirements.Within one week, you’ll receive an e-mail detailing the required amendments. 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