Stunting prevalence is an indicator of a country’s progress towards United Nations’ Sustainable Development Goal 2, which is to end hunger and achieve improved nutrition. Accelerating progress towards reducing stunting requires a deeper understanding of the factors that contribute to linear growth faltering. We conducted path analyses of factors associated with 18-month length-for-age z-score (LAZ) in four prospective cohorts of children who participated in trials conducted as part of the International Lipid-Based Nutrient Supplements Project in Ghana (n=1039), Malawi (n=684 and 1504) and Burkina Faso (n=2619). In two cohorts, women were enrolled during pregnancy. In two other cohorts, infants were enrolled at 6 or 9 months. We examined the association of 42 indicators of environmental, maternal, caregiving and child factors with 18-month LAZ. Using structural equation modelling, we examined direct and indirect associations through hypothesised mediators in each cohort. Out of 42 indicators, 2 were associated with 18-month LAZ in three or four cohorts: maternal height and body mass index (BMI). Six factors were associated with 18-month LAZ in two cohorts: length for gestational age z-score (LGAZ) at birth, pregnancy duration, improved household water, child dietary diversity, diarrhoea incidence and 6-month or 9-month haemoglobin concentration. Direct associations were more prevalent than indirect associations, but 30%–62% of the associations of maternal height and BMI with 18-month LAZ were mediated by LGAZ at birth. Factors that were not associated with LAZ were maternal iron status, illness and inflammation during pregnancy, maternal stress and depression, exclusive breast feeding during 6 months post partum, feeding frequency and child fever, malaria and acute respiratory infections. These findings may help in identifying interventions to accelerate progress towards reducing stunting; however, much of the variance in linear growth status remained unaccounted for by these 42 individual-level factors, suggesting that community-level changes may be needed to achieve substantial progress. Previous cross-sectional studies have identified key risk factors for stunted growth, including being small for gestational age at birth, poor sanitation, childhood diarrhoea, and low dietary diversity and infrequent infant feeding.We found consistent associations of 18-month length-for-age z-scores (LAZ) with maternal height and maternal body mass index.The factors with the strongest associations with 18-month LAZ were length for gestational age z-score (LGAZ) at birth, maternal height and gestational age at birth.Other factors that showed significant associations with 18-month LAZ , were improved household water source, child dietary diversity, childhood diarrhoea incidence and 6-month or 9-month haemoglobin concentration.Interventions targeting these factors associated with LAZ may accelerate progress towards reducing stunting; however, much of the variance in linear growth status remained unaccounted for by individual-level factors, suggesting that community-level changes may be needed to achieve substantial progress and further research is needed to understand the causes of stunting.Linear growth faltering during early life is associated with later health conditions, such as cardiometabolic disease,Several previous reviews have summarised these risk factors,=5846) who participated in the International Lipid-Based Nutrient Supplements (iLiNS) Project in Ghana, Malawi and Burkina Faso. The first objective was to identify the factors associated with 18-month linear growth status, reflecting cumulative linear growth during gestation and the first 18 months after birth, in each iLiNS cohort. The second objective was to identify the pathways through which these factors operate. The prospective design enabled examination of factors associated with linear growth throughout most of the first 1000 days after conception. Harmonisation of many of the variables across cohorts allowed examination of the same factors and pathways in different contexts.In the current study, we performed a set of path analyses of factors associated with linear growth status at age 18 months in four longitudinal cohorts of children  across cohorts accounted for more variance in LAZ than the within-cohort models. Evidence suggests that linear growth faltering is a whole-population phenomenon. In populations with a high prevalence of stunting, it is not the case that a subgroup of that population is faltering while the rest are growing normally, but instead there is usually a downward shift in the entire height distribution of the population.Based on several previous frameworks,At an individual level, maternal height may be related to the child’s genetic potential for adult height that can be attained. In populations with a high prevalence of stunting, maternal height is also partly a reflection of growth restriction experienced by the mother during early life. Therefore, inclusion of maternal height in the model served two purposes: first, to adjust for a proxy of genetic potential, and second, to test the pathway that intergenerational effects of maternal growth during early life, reflected by maternal adult height, on child linear growth may be mediated by (1.1) socioeconomic conditions of the current generation, (1.2) maternal adult factors , (1.3) caregiving practices, (1.4) child factors or (1.5) may directly affect linear growth .Socioeconomic disparities and other environmental effects on child linear growth may be mediated by (2.1) maternal factors, (2.2) caregiving practices, (2.3) child factors or (2.4) may directly affect linear growth.Effects of maternal factors on child growth may be mediated by (3.1) child factors, (3.2) caregiving practices or (3.3) may directly affect child growth.Effects of infant feeding practices on child growth may be mediated by child factors (4.1) or caregiving practices may directly affect child growth (4.2).Effects of preterm birth or intrauterine growth restriction on later child linear growth status may be mediated by (5.1) postnatal child factors /iron status, physical activity, stress) or (5.2) may be direct effects.Effects of child factors  on child growth (6.1) may be mediated by caregiving behaviour in response to these factors, or (6.2) may be direct effects.In the iLiNS-DYAD-G trial in Ghana (n=1320) and the iLiNS-DYAD-M trial in Malawi (n=869), pregnant women were enrolled at ≤20 weeks of gestation. In the iLiNS-DOSE trial in Malawi (n=1932) and iLiNS-ZINC trial in Burkina Faso (n=3220), infants were enrolled at age 6 and 9 months, respectively. All participants were assigned to receive various doses and formulations of lipid-based nutrient supplements (LNS), or to control groups until age 18 months, when length was measured.In the path analyses reported here, we included all children for whom LAZ at age 18 months was available, comprising 1039 children in iLiNS-DYAD-G, 684 in iLiNS-DYAD-M, 1504 in iLiNS-DOSE and 2619 children in iLiNS-ZINC.Detailed reports of the data collection procedures in each trial have been published elsewhere10.1136/bmjgh-2018-001155.supp1Supplementary dataData on socio-demographic characteristics and maternal anthropometric status were gathered at enrolment. In the two DYAD trials, maternal pre-pregnancy body mass index (BMI) was estimated based on BMI and gestational age at enrolment. Capillary or venous blood samples were collected from mothers and/or children at multiple time points for the assessment of (1) malaria using a rapid diagnostic test, (2) Hb concentration (g/L), (3) biomarkers of iron status, including zinc protoporphyrin (ZPP) concentration (μmol/mol heme), and soluble transferrin receptor , and (4) biomarkers of inflammation, including alpha-1-acid glycoprotein  concentration. In ZINC, known HIV infection was an exclusion criterion, but HIV was not tested; therefore, HIV status of women who were enrolled was unknown. In DOSE, HIV status was also unknown. In DYAD-G, HIV infection was known from antenatal cards and HIV-positive women were excluded. In DYAD-M, women were tested for HIV at enrolment but were not excluded.Maternal and/or child saliva samples in DYAD-M and DYAD-G were collected at several time points for the measurement of cortisol concentration (nmol/L). Maternal self-reported stress was measured in DYAD-M at multiple time points using the Perceived Stress Scale.Children were visited weekly for morbidity surveillance. At these visits, caregivers were asked whether the child experienced any illness symptoms, including fever, diarrhoea, vomiting, cough, nasal discharge, respiratory distress or poor appetite during the past seven days and/or data collectors measured the child’s auricular temperature. Longitudinal prevalence and/or incidence of diarrhoea, fever, malaria, acute respiratory infection and/or poor appetite were calculated . In DOSEInfant feeding practices were assessed at multiple time points through qualitative 24 hours and/or 7-day dietary recall questionnaires.In DYAD-G and DYAD-M, gestational age at enrolment was mainly determined by ultrasound and this was used to calculate gestational age at birth. In DYAD-G, infant weight and length were measured within 48 hours of birth or between 3 and 14 days after birth for 87 (9.4%) when the former was not possible. In DYAD-M, infant weight and length were measured within 6 weeks of birth. We estimated length and weight at birth based on LAZ and WAZ measured within 6 weeks of birth, assuming that LAZ and WAZ did not change from birth to the time of measurement. Length and weight for gestational age at birth z-scores were then calculated based on the INTERGROWTH-21st standards.We examined the distribution of each independent variable separately by cohort. We log-transformed skewed variables and winsorised outliers to the 1st and 99th percentile. If transformation did not result in a normal distribution, we created a binary variable. All continuous variables were standardised to SD units by subtracting the mean and dividing by SD.We performed exploratory mediation analyses according to the following steps. We refer to c in a. The third condition was that M is associated with Y, represented by b. In this first step, we tested the significance of b and c.The first condition for inclusion in our mediation analysis was that X is associated with Y, represented by r>0.6), we dropped the one that was less strongly associated with 18-month LAZ. Third, we examined four multivariate models with each category of factors together predicting 18-month LAZ and dropped any that were not associated at p<0.05. If, at this step, no variable was significantly associated when controlling for the others, we retained the one that was most strongly associated with 18-month LAZ. The four multivariate models were (1) all environmental factors together, (2) all maternal factors together, excluding maternal height, (3) all caregiving factors together and (4) all child factors together, excluding gestational age at birth and LGAZ at birth.Second, we reduced the number of variables measuring the same construct by eliminating variables that were collinear. If two variables were highly collinear , child sex and child age at 18-month LAZ assessment. In the final models, we corrected p values for multiple comparisons using the Benjamini-Hochberg correction,Finally, we ran the final path model using the 2 in the models with these 16 predictors in each cohort separately and in the pooled model to determine whether the pooled analysis accounted for more variance in 18-month LAZ than the within-cohort models.For objective 3, we examined the 16 variables that were available in all four cohorts: household asset index, household food insecurity access index, maternal and paternal education, household water and toilet, maternal age, height, and BMI, child diarrhoea and fever prevalence, child 6-month or 9-month Hb and ZPP, mean dietary diversity across time points, variety of play materials and activities with caregivers at age 18 months. We report the RSummary statistics for all independent variables are presented in Among environmental variables, 18-month LAZ was positively associated with higher household assets (DYAD-G), household being located closer to the market (DOSE), higher maternal education , higher paternal education (DYAD-M) and an improved water source . Of the maternal variables, we observed higher LAZ among children of mothers who were taller in all four cohorts and among children of mothers who had higher BMI in three cohorts . In DYAD-G, higher LAZ was associated with higher maternal Hb at ≤20 weeks gestation. Among the caregiving variables, we observed higher LAZ among children who had greater dietary diversity (at 15 months in DOSE and at 9 months in ZINC), higher 18-month variety of play materials  or higher 18-month activities with caregivers (DYAD-M).Of the child variables, higher LAZ was associated with higher Hb at 6  or 9 months (ZINC), greater positive change in Hb from 9 to 18 months (ZINC), lower 18-month AGP (DYAD-M), lower diarrhoea incidence from 6 (DOSE) or 9 (ZINC) to 18 months, lower prevalence of poor appetite from 6 to 18 months (DYAD-G) and higher activity levels as measured by accelerometer mean vector magnitude at 18 months (DOSE). In the two cohorts enrolled before birth, both gestational age at birth and LGAZ at birth were strongly associated with 18-month LAZ. All coefficients in the final models are presented in Regarding pathway (1), examination of genetic effects or intergenerational effects of maternal growth during early life, reflected by maternal adult height, on child 18-month LAZ consistently showed that this was largely a direct association  rather than via mediation by other factors. In the two cohorts in which length at birth was measured, LGAZ at birth mediated a substantial proportion of this association . A very small proportion of this association was mediated by child Hb at 6 months (2% in DOSE) or dietary diversity in infant feeding (1% in ZINC).For pathway (2), the extent to which socioeconomic disparities were mediated by other factors varied across cohorts and independent variables, from 0% to 43%. In ZINC, the association of maternal education with 18-month LAZ was mediated by the child’s variety of play materials (43%) and the association of improved water source with LAZ was not mediated by any other variables. In DYAD-G, the association between household assets and child LAZ was partly mediated by maternal lower Hb concentration in early pregnancy (17%) and gestational age at birth (11%). In DOSE, three environmental variables were associated with LAZ: distance to the nearest market, maternal education and an improved water source. The direct proportion of these associations ranged from 74% to 93%. The association of distance to the nearest market with LAZ was partly mediated by maternal BMI (11%) and child diarrhoea incidence (5%). The association of maternal education with 18-month LAZ was partly mediated by maternal BMI (7%). The association of improved water source with child LAZ was partly mediated by child physical activity (26%). In DYAD-M, while the association of paternal education with LAZ was retained in the variable selection process, it was not significant in the final model and was not mediated by any variables.Regarding pathway (3), the extent to which associations of maternal health and nutritional status with child LAZ were mediated versus direct effects also varied by cohort and independent variable (0%–56%). In DYAD-M, the association between maternal BMI and 18-month LAZ was largely mediated by LGAZ at birth (62%). In the two cohorts in which LGAZ at birth was not available, the association of maternal BMI with LAZ was largely a direct association (92%–94%), with a small proportion mediated by the child’s variety of play materials (8% in ZINC) or child Hb concentration at 6 months (6% in DOSE). In DYAD-G, the association of maternal Hb in early pregnancy with 18-month LAZ was not mediated by any other variables.Associations of dietary diversity in infant feeding at 9 months in ZINC and 15 months in DOSE with 18-month LAZ (pathway 4.1) were partly mediated by child diarrhoea incidence. Caregivers who reported higher dietary diversity in infant feeding also reported higher diarrhoea incidence in the child. The association of dietary diversity with 18-month LAZ was also positive and the association of diarrhoea incidence with 18-month LAZ was negative; therefore, the proportion of the association between dietary diversity and child LAZ that was mediated by diarrhoea incidence was negative (−5% in ZINC and −28% in DOSE).Results from both trials that measured gestational age and length at birth showed that these were direct associations, unmediated by any other child factors (pathway 5.2). The following results from the ZINC cohort supported pathway (6.1): children with lower Hb at age 9 months were provided with lower variety of play materials at 18 months, mediating 12% of the association with 18-month LAZ. Similar to dietary diversity, variety of play materials was positively associated with diarrhoea incidence, therefore this mediated −8% of the negative association between diarrhoea incidence and child LAZ.The final models accounted for 34% of the variance in 18-month LAZ in DYAD-G, 37% in DYAD-M, 17% in DOSE and 20% in ZINC. Overall, maternal and child factors showed stronger associations with 18-month LAZ compared with environmental and caregiving factors, even when considering both indirect and direct effects .For objective 3, the 16 variables common across all four cohorts accounted for the following percentage of variance in 18-month LAZ: 16% in DYAD-G, 21% in DYAD-M, 15% in DOSE, 19% in ZINC, and 25% in the pooled analysis. Of these 16 variables, maternal height accounted for the largest proportion of variance in 18-month LAZ: 11% in DYAD-G, 14% in DYAD-M, 10% in DOSE, 11% in ZINC and 9% in the pooled analysis. Therefore, excluding maternal height, environmental, maternal, caregiving and child factors accounted for only 5%–8% of variance in LAZ in the individual cohorts and 16% in the pooled analysis.In four prospective cohorts of children, we conducted path analyses of environmental, maternal, caregiving and child factors hypothesised to be associated with 18-month linear growth status. Out of 42 indicators examined, only 2 indicators emerged as consistent predictors of 18-month LAZ in 3–4 cohorts: maternal height and maternal BMI. Two additional factors, which were only measured in the two pregnancy cohorts, were significantly associated with 18-month LAZ in both cohorts in which they were measured: LGAZ at birth and gestational age at birth. In the two cohorts enrolled after birth, four factors were significantly associated with 18-month LAZ in both cohorts: improved household water source, dietary diversity in infant feeding, childhood diarrhoea incidence and 6-month or 9-month Hb. Overall, maternal and child factors showed stronger and more consistent associations with child LAZ than environmental and caregiving factors, even when considering both indirect and direct effects.While specific pathways differed across cohorts, in general direct associations were stronger and more prevalent than indirect associations. The strongest evidence for mediation was the finding that 62% of the association of maternal BMI with 18-month LAZ was mediated by LGAZ at birth in DYAD-M. This underscores the importance of maternal nutritional status for fetal growth, with lasting consequences for growth status later in childhood. LGAZ at birth also mediated a substantial proportion of the association of maternal height with 18-month LAZ (30%–34%) in both cohorts in which it was measured (DYAD-M and DYAD-G). This pathway is likely to reflect genetic influences, at least in part, and therefore may be only partly modifiable. In the two cohorts in which LAZ at birth was not measured (DOSE and ZINC), the association of maternal height with 18-month LAZ was a direct association.et al used population surveys to estimate cases of stunting attributable to 18 risk factors in 137 developing countries, including indicators of maternal nutrition and infection, teenage motherhood and short birth intervals, fetal growth restriction and preterm birth, child nutrition and infection, and environmental factors. They found that a large number of stunting cases were attributable to being born small for gestational age at term , followed by poor sanitation  and diarrhoea .et al calculated the population-attributable fraction of stunting for five categories of risk factors: low birth weight, insufficient diet, enteric dysfunction, infection and toxin exposure. They estimated that 25% of global stunting could be attributed to child diarrhoea incidence. In Africa, the largest number of stunting cases was attributable to low dietary diversity, low feeding frequency, or both (30%).Of the eight key risk factors closely associated with linear growth faltering that we found, three were consistently identified in previous studies that examined the relative contribution of risk factors to stunting prevalence globally: fetal growth restriction, diarrhoea and low dietary diversity. Danaei The key risk factors for linear growth faltering that we identified were similar to those identified in these studies despite several differences in methodology, including prospective cohorts rather than cross-sectional design, examination of both direct and indirect associations through path analysis, and examination of LAZ as a continuous variable rather than stunting cases. By analysing associations with LAZ, we were able to estimate the difference in LAZ with each SD change in predictor variables. Each SD difference in maternal height (5–6 cm) was associated with 0.25 difference in LGAZ at birth (0.5 cm) and a 0.2–0.3 difference in child 18-month LAZ (0.5–0.8 cm). Each SD difference in LGAZ at birth (1.9 cm) was associated with about a 0.5 difference in child 18-month LAZ (1.4 cm). Each SD difference in gestational age at birth (1.8 weeks) was associated with a 0.13–0.17 difference in 18-month LAZ (0.3–0.5 cm). The coefficient sizes for the other key risk factors  were smaller in magnitude (0.06–0.12).Despite the examination of 42 environmental, maternal, caregiving and child factors measured longitudinally throughout most of the first 1000 days, our models accounted for a relatively small proportion of the variance in 18-month LAZ (17%–37%), and a large proportion of that was due to maternal height (10%–14%). Factors that were not associated with LAZ were indicators of maternal iron status, illness and inflammation during pregnancy, maternal stress, depression and cognition, exclusive breast feeding during 6 months post partum, infant feeding frequency and child fever, malaria and acute respiratory infections. Variance that remained unaccounted for could be due to unmeasured factors, such as enteric dysfunction, asymptomatic infections, microbiome composition, maternal HIV status or aspects of the diet that were not assessed. It is also possible that population-level factors, such as community sanitation or access to healthcare, may be stronger drivers of growth stunting compared with individual or household-level factors.Our finding that the pooled analysis accounted for more variance in LAZ (25%) compared with separate analyses by cohort (11%–21%) also supports this argument, although this difference is not very large. Excluding the variance accounted for by maternal height, the pooled analysis accounted for two to three times the variance in LAZ (16%) compared with the separate cohort models (5%–8%). This may be due to the lack of inclusion of a cohort experiencing healthy growth. All four iLiNS cohorts experienced growth faltering, although the cohorts in Malawi and Burkina Faso had faltered 1.5–1.9 SD below the WHO median on average, compared with the Ghanaian children whose mean LAZ was only 0.8 SD below the WHO median by age 18 months. Another potential reason is that we did not measure community-level or social factors that may contribute to children’s growth faltering at a population level. For example, several studies have found that community-level sanitation coverage predicts child growth more strongly than household-level sanitation.Strengths of the study were the large number of children in each cohort, the large number of variables examined and the availability of data from four cohorts of children in three African countries, with many variables harmonised across cohorts, enabling the examination of the same factors and pathways across contexts. Another strength was that the cohorts were enrolled either during pregnancy or at 6 or 9 months post partum and followed prospectively through 18 months post partum, allowing these risk factors to be examined throughout a large portion of the first 1000 days after conception. The prospective design allowed associations in the path model to be drawn from variables measured at earlier time points  to variables measured at later time points , reducing the risk of reverse causality. However, this does not prevent the possibility of residual confounding, that is, unmeasured variables accounting for observed associations, therefore we are not able to establish causality. Another limitation was that we were not able to explore certain risk factors, such as asymptomatic infections or enteric dysfunction. We reported multiple comparisons; therefore, may have found some false positive associations due to chance. However, we reduced this possibility by reporting p values corrected for multiple comparisons and by basing our primary conclusions on findings that were replicated in more than one cohort. These samples may not be representative of their populations due to their participation in the nutritional supplementation trials. However, any bias introduced by the supplementation would be expected to decrease rather than increase the strength of the associations.Our findings may help to inform the design of interventions to pursue United Nations’ Sustainable Development Goal 2, to end hunger and achieve improved nutrition by 2030, with the reduction of stunting prevalence as a key indicator of progress towards this goal. The evidence we present shows that child linear growth status is closely linked to higher maternal height and BMI, fetal growth, full-term birth, improved household water supply, higher child dietary diversity, Hb during infancy and reduced childhood diarrhoea incidence, but we did not find associations with maternal iron status, illness and inflammation during pregnancy , maternal stress and depression, exclusive breastfeeding during 6 months post partum, infant feeding frequency or child fever, malaria and acute respiratory infections. Much of the variance in linear growth status remained unaccounted for by the 42 individual-level factors measured in this study, suggesting that interventions targeting other individual-level factors or community-level changes may be needed to achieve substantial progress in stunting reduction. The large amount of unexplained variance also suggests that further research is needed to understand the causes of linear growth faltering in order to design more effective strategies to reach the WHO’s target to achieve a 40% reduction in the number of children under-5 who are stunted by 2025.


Intramural esophageal bronchogenic cyst is very rare. Surgical removal of the cysts is advised even the patients are asymptomatic, since the cyst can lead to complications, and there is a risk of malignant transformation. Thoracotomy or thoracoscopy is the most commonly used approach for complete excision of the cysts. To our knowledge, this is the first report to excise intramural esophageal bronchogenic cyst completely by endoscopic submucosal tunnel dissection (ESTD).A 40-year-old male was referred to our hospital due to the detection of a submucosal tumor at the distal esophagus. The tumor was found during gastroendoscopy in a general health check-up. The patient had no symptoms. A benign esophageal tumor was confirmed by endoscopic ultrasonography (EUS) and computed tomography (CT). On the basis of these results, ESTD was performed. During the procedure, a cystic mass was observed between the mucosa and the muscular layers of the esophagus, and a hybrid knife was used for dissection. Histopathological examination showed the cyst wall was lined by pseudostratified ciliated columnar epithelium, consistent with a bronchogenic cyst. The esophagography using meglumine diatrizoate showed no leakage on the seventh day after ESTD. The patient remained asymptomatic and had a regular diet during the follow-up period.We successfully utilized ESTD for complete removal of esophageal bronchogenic cysts originating from the muscularis propria. The approach appeared safe, providing a minimally invasive treatment option for patients. Bronchogenic cyst is a congenital lesion derived from the primitive foregut, and is one of the most common primary cysts of the mediastinum. Although periesophageal bronchogenic cysts have been frequently reported, a completely intramural cyst is rare . BronchoA 40-year-old male was referred to our hospital due to the detection of a submucosal tumor at the distal esophagus. The tumor was found during gastroendoscopy in a general health check-up. The patient had no dysphagia, chest pain, vomit, fever, cough nor dyspnea. No clinically significant abnormalities were found during physical examination. Laboratory results were all within normal range, which included serum tumor marker , complete blood test, erythrocyte sedimentation rate, hepatic and renal function tests. During the gastroendoscopy, an abnormal swelling covered with normal mucosa was found on the right wall of esophagus at 35 cm from the incisors Fig. . The patBronchogenic cyst is a benign congenital malformation of the mediatisnum, and the cyst can be found in a variety of locations such as paratracheal, carinal, or hilar. Esophageal brochogenic cyst is extremely rare , 4. MostSurgical resection of bronchogenic cyst is often recommended, as the cyst can lead to complications including infection, hemorrhage, rupture and malignant degeneration if left untreated . To dateTo our knowledge, this is the first report that utilized ESTD to excise the esophageal bronchogenic cyst completely. Tang et al. reported a treatment of esophageal brochogenic cyst with ESTD, the cyst was not completely excised by ESTD . During The use of ESTD allows en bloc resection. A previous report indicated a recurrence of mediastinal cysts 15 years after incomplete resection ; therefoIn our experiences, EUS is sensitive for distinguishing cystic from solid masses. It facilitates the diagnosis of intramural esophageal lesions by providing clear images delineating the size, layer of origin, and location of the cyst relative to esophagus. In summary, we reported the successful use of ESTD to completely remove the esophageal bronchogenic cyst originated from the muscularis propria. The procedure was minimally invasive and appeared safe, although a long-term follow-up for further assessment of the long-term outcome is needed.


Adverse birth outcome which attributes to most perinatal deaths is an important indicator of child health and survival. Hence, this study aims to identify determinants of adverse birth outcome among mothers who gave birth in public hospitals of Tigrai region, North Ethiopia.P value < 0.05.Hospital based case-control study was conducted in Tigrai region, Ethiopia between December 2015 and January 2016 among 405 (135 cases and 270 controls) consecutively selected mothers who gave birth in four randomly selected public Hospitals. Mothers with adverse birth outcome  were the cases while mothers without adverse birth outcome  were the controls. Data were collected by interview and reviewing medical records using structured questionnaire. The collected data were entered into database using EPI info version 3.5.1 then exported to SPSS version 21 for analysis. Finally, multivariate logistic regression was used to identify determinants of adverse birth outcomes at The mean age of cases and controls was 27.3 (SD = 6.6) and 26.14 (SD = 4.9) years, respectively. In a multivariate analysis; less than four antenatal care visits , not receiving dietary counseling , not using family planning methods , less than 24 months inter pregnancy interval , and less than 11 g/dl hemoglobin level  were significantly associated with adverse birth outcomes.The number of antenatal care visits, ever use of family planning methods, not receiving dietary counseling during antenatal care follow up visits, short inter-pregnancy interval, and low hemoglobin level were identified as independent determinants of adverse birth outcome. A concerted effort should be taken improve family planning use, and antenatal care follow-up with special emphasis to maternal nutrition to prevent adverse birth outcomes. Adverse birth outcome defined as low birth weight (LBW); birth weight < 2.5 kg, preterm birth (PTB); < 37 gestational weeks at birth) and stillbirth  continues to be public health problem globally mainly in developing countries –8. In 20Though adverse birth outcomes and their consequences are preventable , 12, 13,Different studies revealed that different factors could contribute for adverse birth outcome; for instance socio-demographic characteristics such as residence , age, maTo investigate the problem, epidemiological data pertaining to determinants of adverse birth outcome are required for planning and implementing proper interventions. However, determinants of adverse birth outcomes are poorly documented and understood in the study area. Therefore, local epidemiological studies are fundamental, to assess determinants of adverse birth outcome and identify target areas for future interventions. Thus, this hospital based case control study aims to identify determinants of adverse birth outcome among mothers who gave birth in public hospitals, Tigrai region, North Ethiopia.Hospital based unmatched case-control study was conducted from December 2015 to January 2016 among mothers who gave birth in four randomly selected public hospitals to identify determinants of adverse birth outcome. The study was conducted in Tigrai region, northern Ethiopia. The region is administratively divided into 7 zones, 52 woredas and 814 Kebeles . In Tigrai regional state, there are 2 comprehensive specialized hospitals, 15 general hospitals, 20 primary hospitals, 204 health centers, 712 health posts and over 500 private health facilities. Our study was conducted in general hospitals. General hospitals provide inpatient and outpatient services, including maternal and child care. General hospitals also serve as a referral center for primary hospitals. Hence the setups of the general hospitals are similar in kind of services they deliver, health information recording system, and in number and qualification of health professional required.According to the 2007 census projection, the total population of Tigrai regional state is 5,055,999 . A total estimated number of 116,506 births have been registered annually in the region. For this particular study, the source population was all mothers who gave birth and infants in pair in general hospitals in Tigrai region, Northern Ethiopia. Cases were mothers with adverse birth outcome , and controls were mothers without adverse birth outcome .The sample size was determined using EPI-Info 3.5.1 statistical software. The assumptions for the sample size calculation were: proportion of non-educated mothers among controls , minimumData were collected by interview using structured questionnaire adopted from Ethiopian Demographic and Health Survey and other related literatures , 30–32. The weight of the newborns was measured using a Salter’s hang-up scale to the nearest 100 g following standard techniques within 15 min after delivery and basic newborn care has been given. The newborns were weighted naked or in minimal clothing. Maternal MUAC (mid upper arm circumference) was also measured following the appropriate procedure to the nearest centimeter using the standard measuring tape.p < 0.25 in the bivariate analyses as candidates, multivariable logistic regression model was fitted to identify independent determinants of adverse birth outcome. Finally, variables at a P < 0.05 were considered statistically significant in the final multivariable logistic regression model. Model fitness was checked by Hosmer and Lemeshow . Adjusted odds ratio (AOR) with 95% CI was computed to determine the strength of association between the variables of interest.Data were coded, cleaned, edited and entered to EPI info version 3.5.1 and exported to SPSS version 21 for analysis. Descriptive statistics such as means and proportions were used to summarize the data. In this study, cases were coded as 1 and controls were coded as 0. Bivariate logistic regression analysis was used to see the unadjusted effect of each independent variable on the dependent variable of the study. Considering all independent variables with a A total of 405 mothers (135 cases and 270 controls) were included in the present study. The cases were consists of low birth weight, preterm and stillbirth; 78 (57.8%), 45 (33.3%) and 12 (8.9%) respectively. The mean age of cases and controls with standard deviation (SD) was 27.3 ± 6.6 and 26.14 ± 4.9 years, respectively. Majority of mothers, 84.4% of the cases and 93.0% of the controls were married. Among the total mothers included in the study, 46 (34.1%) of the cases and 37 (13.7%) of the controls were unable to read and write. Most mothers, 72.6% of the cases and 70% of the controls were housewives. Likewise, majority of the cases (89.6%) and the controls (88.9%) were Orthodox Christian by religion. Seventy seven (57%) of the cases and 221 (81.5%) of the controls were living in urban  of cases and 221 (81.8%) of controls had less than four parity. Among participants of the study, 61 (70.1%) cases and 40 (27.6%) controls had less than 24 months inter-pregnancy interval. Majority of the participants of the study, 110 (81.5%) cases and 270 (100%) controls attended ANC at least once in health facility. Meanwhile, mothers who had at least four ANC visits were 40 (29.6%) among cases and 195 (72.2%) among controls. Among the total participants 88 (65.2%) of the cases and 30 (11.1%) of the controls did not receive dietary counseling during the index pregnancy. Among the total mothers included in the present study, 76 (56.3%) of the cases and 235 (87%) of the controls had hemoglobin level of 11 g/dl or more during the index pregnancy. In addition, in 50 (37%) of the cases and 28 (10.4%) of the controls the index pregnancy was unwanted , receiving dietary counseling during the prenatal care of the index pregnancy, ever use of family planning, wanted/plan of the index pregnancy, experience complication in the index pregnancy, and hemoglobin level.After controlling for confounders using multivariate analysis, number of ANC visit, receiving dietary counseling or not during the index pregnancy, inter-pregnancy interval, ever use of family planning methods, and hemoglobin level were identified as independent determinant factors of adverse birth outcome. That is; mothers who had less than four ANC visits were about four times more at risk of experiencing adverse birth outcome than mothers who had at least four ANC visit . Mothers who did not receive dietary counseling were 11 times more at risk of experiencing adverse birth outcomes compared to their counterparts, . Furthermore, mothers with hemoglobin level of less than 11 g/dl were nearly five times more at risk to experience adverse birth outcome than those with hemoglobin level of 11 g/dl or more . Ever use of family planning methods was also another important variable independently associated with adverse birth outcome; mothers who had not ever used family planning methods were four times more likely to experience adverse birth outcome compared to their counterparts, . Mothers with less than 24 months inter pregnancy interval were five times more risk to experience adverse birth outcome than mothers with greater than 24 months inter-pregnancy interval, [AOR = 5.21, 95% CI: 1.89–13.86  inter-pregnancy interval were five times more likely to experience adverse birth outcome compared to their counterparts. This result is in line with studies in Indonesia, India, Tanzania and Ethiopia , 43–46, This study has its own drawbacks. The study did not assess the contribution of important variables such as knowledge, husband involvement and food security. Furthermore, this study is not free from recall and social desirability biases.In this study less than four ANC visits, not receiving dietary counseling during the index pregnancy, not ever using family planning methods, inter-pregnancy interval of less than 24 months, and hemoglobin level of less than 11 g/dl were significantly associated with adverse birth outcomes. Therefore, a concerted effort should be taken to improve family planning, and ANC follow-up with special emphasis to maternal nutrition to prevent adverse birth outcome.


Fetal growth has been known to be associated with particulate matter (PM) air pollution during gestation. Given that regular working may deviate outdoor air pollution exposure, the association between air pollution and fetal growth restriction can be different across maternal working status. This study was to assess possible effect modification by maternal employment in the association between exposure to PM during pregnancy and fetal growth restriction.10 and PM2.5) regulatory monitoring data for 2001–2012 and 2008–2012, respectively, and birth certificate data for 2002–2012, we computed maternal exposures with district-level averages of PM10 and PM2.5 during one year before birth, entire pregnancy, and the 1st, 2nd and 3rd trimesters. The outcomes of fetal growth restriction were assessed by small for gestational age  as well as low birth weight  at term. We performed logistic regression to examine the association between PM and each of fetal growth restriction outcomes adjusting for individual risk factors. For effect modification by maternal employment, we estimated adjusted odds ratio (OR) of SGA or LBW for interquartile (IQR) increases in PM10 or PM2.5 stratified by employed and non-employed mothers. We also computed relative excess risk due to interaction (RERI) to investigate additive interaction.Using hourly PM less than or equal to 10 and 2.5 μm in diameter (PMP < 0.001). SGA occurred in 12.7% of employed and 12.8% of non- employed (P = 0.124) mothers. For non-employed mothers, we observed increased odds of SGA per IQR increase in PM10 for one year before birth : 1.00–1.04, P = 0.028). ORs of SGA for full pregnancy period and the 3rd trimester were also positive but did not reach statistical significance. We did not observe positive association for PM2.5. RERI was not significant both for PM10 and PM2.5.Among 824,011 singleton term births, 34.0%  were employed and 66.0%  were non-employed mothers. Proportions of LBW were 1.5% in employed and 1.6% in non-employed  contains supplementary material, which is available to authorized users. Fetal growth is assessed by comparing birthweight of the newborn with expected weight for the baby’s gestational age. Specifically, low birth weight   and smalFetal growth restriction is mostly caused by utero-placental dysfunction leading to inadequate supply of nutrients and oxygen to support normal growth of the fetus , 7. FemaAmong the physical environment, particulate matter (PM) air pollution is recognized as a risk factor for preterm birth and fetal grown disorders . Previou10 and PM2.5) exposed during the entire period of pregnancy which was not evident for the 1st or 2nd trimester  were defined as SGA. Gestational age is based on the physician’s final estimate of gestation using ultrasound taken early in pregnancy and the mother’s date of the last menstrual period. SGA was referred in weeks, rounding off to the nearest completed week [ were def10 and PM2.5 concentrations measured at maximum 40 regulatory air pollution monitoring sites in Seoul during 2001–2012 from the National Institute of Environmental Research as described in previous studies [2, at least one urban background site is located in each of the 25 districts . We did not included two additional sites in two gu’s because these sites operated for a few years. Because relatively small numbers (6–20) of monitoring sites measured PM2.5 before 2007, the analysis for PM2.5 was restricted to births between 2008 and 2012.We obtained hourly PM studies , 38. Brigu level, the PM concentration measured at a single monitoring site in each gu was assigned to all mothers residing in the same gu. To compute the five metrics, we calculated the 24-h daily averages for days on which > 75% of hourly measurements (18 h) were available at each site. The application of this inclusion criterion for day resulted in exclusion of 6.9% of days for 11 years and at 25 monitoring sties. Then, we averaged daily concentrations over each of the following five pregnancy periods: one year before birth, whole pregnancy, and the 1st, 2nd and 3rd trimesters. To use exposure estimates consistently for a fixed time period, the 1st, 2nd and 3rd trimesters were defined as 0–13+ 6, 14+ 0–27+ 6, and 28+ 0–36+ 6 weeks, respectively. To derive representative exposure estimates in each period, we computed one year exposures for mothers using the sites with more than 9 months of daily data and three trimester-exposures for those with at least 75% of the data during each trimester available. All 25 urban background sites met the inclusion criteria for each of the five periods.Using hourly PM measurement data, we computed five exposure metrics corresponding to five antenatal periods at the mothers’ home addresses. Because mothers’ home addresses are available at the We used directed acyclic graphs (DAGs) analysis to examine potential confounders and effect modifiers in the association between air pollution and LBW or SGA  to assess additive interaction [We used chi-square test  and student t-test (for continuous variables) to assess the difference in individual characteristics as well as fetal growth restriction between employed and non-employed mothers. For the association between PM and fetal growth restriction at term (SGA and LBW), we conducted logistic regression using five exposure metrics after adjusting for individual characteristics separately by employed and non-employed mothers. Individual characteristics included birth date (birth year and month), infant sex, maternal education, maternal age, parity (first childbirth or not), birth season, and gestational age. Non-linear association between birth date and term LBW was adjusted by natural cubic spline with 11 degrees of freedom (df) (1 df per year). Odds ratio (OR) for each birth outcome was estimated per interquartile range (IQR) increase in PMeraction .gu-specific exposures with those of spatially constant exposures, based on daily average PM concentrations over Seoul. Second, we applied mixed models to account for correlation of fatal growth restriction within each of the eight district groups (downtown and areas 1 to 7) and examined the robustness of primary results. These eight district groups were classified based on geographical proximity of 25 districts [Our additional analyses investigated the sensitivity and variation of our primary results. First, we investigated whether the difference in the association between employed and non-employed mothers is enlarged, when exposure assessment is spatially coarse as frequently used in previous studies of air pollution and birth outcomes , 42. Foristricts . The anaThis study was reviewed and approved by the Institutional Review Board of Seoul National University (IRB No. E1503/002–001).10 concentrations of mothers during one year prior to birth, full pregnancy, the 1st, 2nd, and 3rd trimesters were 57.8 (standard deviation = 9.9), 57.3 (10.8), 58.2 (16.5), 57.2 (16.2), and 56.6 (16.6) μg/m3, respectively. PM2.5 concentrations for births in 2008–2012 were 29.1 (6.7), 28.9 (7.0), 29.3 (8.7), 28.8 (8.5), and 28.4 (8.5) μg/m3. PM10 concentrations for 1 year prior to birth were moderately correlated with those during the three trimesters (0.44 to 0.62), whereas the correlations of PM10 concentrations between trimesters were weak (− 0.09 to 0.26)  and outcome  with adjustment for covariates.In our study population of 824,011 singleton term births, SGA and LBW comprised 11.2%  and 1.5% , respectively. The mean PMβ = − 0.005, P < 0.001). The proportion of employed mothers increased from 23.6% in 2002 to 44.7% in 2012. Whereas percentage of LBW increased from 1.4% in 2002 to 1.7% in 2012 (P for trend < 0.001 for both), percentage of SGA was consistent over time   were employed and 66.0% 544,155) were non-employed (Table 55 were n3 (IQR in the full pregnancy period) increase in PM10 for one year before birth was associated with 2% increase in odds of SGA : 1.00–1.04, P = 0.028). The positive association of PM10 for the full pregnancy period and the 3rd trimester with SGA did not reach statistical significance. RERIs indicating additive interaction by maternal employment did not show statistical significance. ORs of LBW were also higher in non-employed mothers than employed mothers, but RERIs were not significant.ORs of SGA and LBW were generally higher in non-employed mothers than employed mothers Table . ORs of 3 increase of PM10 during one year before birth in non-employed mothers remained significant.The positive association was not found when we used coarse exposure assessment applying citywide Seoul-mean concentration. ORs were lower both for SGA and LBW than those in our primary analysis using district-means and became negative for some exposure periods. Our findings were similar, when we applied mixed models to account for additional variability at the district group level  or province was 8–25% during the same period of our study period (2003–2012) [There are other several limitations in this study. First, there is possible information bias because we did not verify mothers’ self-reported working status during pregnancy. Some of the mothers classified as employed might have taken maternity leave during pregnancy and some of the unemployed might have left their job just before birth. As a retrospective study of birth registry data, we could not assess the duration of employment during pregnancy for each mother. Future studies based on prospectively collected data could minimize this possible misclassification bias. Second, because we used the administrative data, some key individual confounders related to smoking intensity, previous pregnancy history, and type of LBW  were not available. As there were previous findings of associations between exposure to PM air pollution and smoking intensity, smoking might have confounded air pollution-related health effect . Conside03–2012) . Relocat03–2012) , the imp03–2012) .Exposure to high ambient PM air pollution during one year before birth tended to increase the risk of SGA at term in non-employed mothers living in Seoul, Korea. Despite potential difference due to mothers’ health status, socioeconomic condition, and exposure misclassification related to their employment status, there was no evidence of effect modification by maternal employment in the association between PM concentration and fetal growth restriction. Futures studies using refined metrics would be helpful to confirm this finding.Additional file 1:Table S1. Correlation coefficients between a pair of the five metrics of mothers’ PM10 concentrations corresponding to the five antenatal periods of 824,011 singleton term births in Seoul, Korea, for 2002–2012. Table S2. Odds ratios (OR) and 95% confidence intervals (CI) for small-for-gestational age (SGA) and low birth weight (LBW) at term per interquartile-increase in PM10* from mixed models that account for correlation of fatal growth restriction at the district group, by employment status in 824,011 singleton term births in Seoul, Korea for 2002–2012. Table S3. Odds ratios (ORs) and 95% confidence intervals (CIs) for small-for-gestational age (SGA) and low birth weight (LBW) at term per interquartile-increase in PM2.5* from mixed models that account for correlation of fatal growth restriction at the district group, by employment status in 386,483 singleton term births in Seoul, Korea for 2008–2012. Figure S1. Map of urban background and urban roadside sites of the regulatory air quality monitoring network in Seoul. Figure S2. Directed acyclic graph of individual characteristics that potentially confound the association between residential PM concentration (“exposure”) and fetal growth restriction (“outcome”). Figure S3. Average concentrations of PM10 and PM2.5 between employed and non-employed mothers by five different antenatal periods of singleton term births in Seoul, Korea. (A) PM10 (B) PM2.5. (DOCX 659 kb)


Measles elimination has been identified as a public health priority in Europe for a long time but has not yet been achieved. The World Health Organization (WHO) recommends identification of susceptible sub-populations to target supplementary immunization activities. We used three different sources of information: retrospective samples investigated for measles IgG between 1997 and 2016, vaccine coverage data from the existing electronic registry for birth cohorts 2015 to 1999, and surveillance data from 2009 until 20 July 2019. We calculated susceptibility by birth cohort using seroprevalence data, adjusting vaccine coverage data with reported effectiveness , and compared it with measles incidence data, aggregated by birth cohorts and districts. Susceptibility levels for persons 10–41 years (birth cohorts 2007–1976) were 10.4% and thus far above the recommended values of WHO (5%). Older birth cohorts were sufficiently protected. Districts with the highest susceptibility estimates corresponded with districts with the highest incidence rates. Birth cohorts with susceptibility levels > 10% showed a 4.7 increased relative risk of having had more than one measles case. We conclude that retrospective serosurveys are a cheap and useful approach in identifying susceptible sub-populations, especially for older birth cohorts whose coverage data remain scarce. Measles elimination in the WHO Region for Europe was already planned for 2010 but has not yet been achieved . SeveralIn Austria, mandatory reporting of suspected measles cases has been implemented in 2006 . The firVaccine coverage is estimated by the administrative method in Austria. Since 2016, an agent-based simulation model has been implemented to better assess the number of susceptible persons to measles .In Tyrol, one of Austria’s nine regions with 739,139 inhabitants in 2016 . Since 2As of 2014, the Tyrolean government has required proof of two doses of MCV from all health care workers (HCW) in Tyrolean hospitals. Those who lack documented evidence of two doses of MMR are required to undergo serological testing. The same applied to all medical students.The WHO provides guidance to seroprevalence surveys to better identify pockets of susceptible populations, and recommends their use if surveillance was only recently implemented and when coverage data are less reliable for older age groups . NeverthIn order to identify sub-populations that are most at risk for measles, we assessed the population immunity using three different sources of information: The results of an analysis of the MCV coverage data extracted from the regional electronic vaccination registry, the results from an ad-hoc retrospective serosurvey conducted in 2017 and the most recent surveillance data from the electronic reporting system EMS [® Anti-Measles Virus/IgG, Siemens, Marburg, Germany) and since then a chemoluminent immunoassay , Italy) were used. A sufficient number of parallel tests were conducted using both kits during the transition phase from ELISA to CLIA and results were comparable between the two kits. The last available laboratory result of each individual obtained between 1 January 1997 and 31 December 2016 was included. The exclusion criteria included the following: (a) place of residency outside Tyrol or not available, (b) date of birth before 1917 or not available, (c) inconclusive laboratory result, or (d) potential immunodeficiency. Potential immunodeficiency was defined as admittance to an intensive care unit (ICU) department or any other department, where patients were likely to be immunocompromised at the time of the sampling, such as oncology, transplant, or hematological wards. Furthermore, we excluded all HCWs and medical students tested for measles IgG from our analysis. Both groups who could not provide documented evidence of vaccination with two doses of MCV were requested by the regional authorities to confirm immunity by IgG testing.We used retrospective data of all persons who were investigated for measles IgG antibodies between 1997 and 2016 in the Division of Virology at the Medical University of Innsbruck . This laThe age of the participants was calculated as of 31 December 2016. Susceptibility to measles was assumed to be unchanged since the time of the last investigation.We defined birth cohorts with susceptibility levels higher than 10% as high risk cohorts.TM . We calculated the susceptibility by birth cohort and district, using the mid population data between 1996 and 2016 [Seroprevalence data were extracted from the virological laboratory software in an anonymized way and analyzed using Epiinfoand 2016 .The study was approved by the ethical committee of the Medical University of Innsbruck under EK Nr: 1082/2017.We extracted the MCV coverage data from the electronic vaccination registry of Tyrol in July 2018 for all available birth cohorts 1997–2013) in an aggregated format . Data fr in an agWe used the case-based measles surveillance reports between 2003 and 2019 from Tyrol  and compBetween 1 January 1997 and 31 December 2016, 24,243 samples were investigated for measles IgG. After application of the exclusion criteria, 13,060 individual records remained and were included in our analysis .n = 7,057); for 1.5% (n = 195) gender information was not available. The highest number of participants was in the age group 40 to 49 years (as of 31 December 2016) (Women accounted for 54% of participants (er 2016) . Childreer 2016) .All nine districts of Tyrol sent samples to the laboratory. The mean proportion of the represented population by district was 1.7% ranging from 0.8% to 2.6%.Among all participants 5.9% were susceptible to measles. Susceptibility levels were age dependent . The higThirty-seven of 99 birth cohorts (excluding the birth 2016 cohort due to non-eligibility for vaccination) did not achieve the necessary immunity levels needed for elimination. Among them, the mean difference between expected WHO susceptibility requirements and observed susceptibility levels was 17.6% . The five most recent birth cohorts (2011–2015) showed levels of 25% or higher above the expected values. Ten birth cohorts  showed susceptibility levels between 5–9%, and 22 birth cohorts  between 1% to 4% deviating from the target, respectively. p < 0.0001). The mean proportion of susceptible persons per district varied between 5.0% and 8.6%, with a mean of 6.4% and a median of 6.0%.Less females were susceptible to measles (5.1%) compared to males (7.1%) (The adjusted susceptibility levels by birth cohort using one (MCV1) or two doses (MCV2) of MCV both followed the pattern of the 3-year moving average seroprevalence susceptibility data, especially for MCV1 for birth cohorts 2009 and older. The mean difference between seroprevalence data and first dose adjusted susceptibility was 5% , and the mean for the second dose was −5% , respectively. While the estimated proportion of susceptible population with one dose of vaccine was below the data from the seroprevalence survey the estimated susceptibility with two doses of MCV was above the seroprevalence estimates—except for the birth cohorts 2010 to 2012 .Between 2003 and 2019, a total of 99 measles cases were reported. Since 2008, peaks with a maximum of 16 cases in 2011 were observed every 3–4 years . Detailen = 34) of cases.Among the 77 cases reported since 2009, the mean age at the time of diagnosis was 20.5 years . Females accounted for 44% . The highest incidence was reported from the district with the highest proportion of susceptible population from the seroprevalence data Among the 100-year birth cohort band, 72 birth cohorts had susceptibility levels below 10% and 61 birth cohorts below 5%. The relative risk of having observed more than one measles case per district between 2009 and 20 July 2019 in districts with susceptibility levels > 10% was 4.7 (95% CI: 2.52–8.52).We assessed the susceptibility levels for measles among the Tyrolean population using three sources of information: a retrospective analysis of already available serological data for measles IgG (serosurvey), vaccine coverage data from the electronic vaccination registry and surveillance data from the mandatory electronic reporting system.We found that the WHO recommended susceptibility thresholds of maximum 15% in children 24 months–4 years, 10% in persons 5–9 years, and 5% in persons >10 years were not achieved. Susceptibility was age dependent and the threshold of less than 5% was only reached by birth cohorts 1976 (41 years of age as of 31 December 2016), 1978, and 1979 or older. Susceptibility levels for persons 10–41 years (birth cohorts 2007–1976) were twice as high as the recommended values of WHO of 5% for these age groups.More than one-third of the population was considered to be immune with susceptibility levels below the requested age-dependent thresholds. The highest susceptibility levels were identified among the most recent birth cohorts until birth cohort 2011, with susceptibility levels differing between 15–50% from the WHO target. Ten birth cohorts revealed susceptibility levels between 5% and 9% and further 22 birth cohorts levels between 1% and 4% above the expected WHO targets. Many member states of the European Union did not achieve the requested WHO goals and susceptible persons accumulated over time . Even a Significantly more males were susceptible to measles compared to females. This uneven gender distribution was also confirmed by surveillance data. Such inequities are not yet addressed through vaccination policies  and studWe identified several limitations of our seroprevalence study: due to the very long observation period, the likelihood of losing detectable antibody levels—especially of persons who were vaccinated with only one dose of MCV—has to be considered and might have biased our results. This would probably not apply for persons who had either experienced natural infection, received two doses of MCV or were re-exposed. This may have been of less influence, as participants from nearly all birth cohorts were investigated recently. In fact, the long time span of the serosurvey resulted in a large sample size and thus allowed more precise estimates. Estimates also became more reliable due to careful exclusion of ambiguous data and they were based on well validated test kits in an accredited laboratory. Despite the fact the only few seroprevalence data of recent birth cohorts were available, the estimation of susceptibles by the two different approaches, retrospective seroprevalence data, and estimation of susceptibles using coverage data corresponded quite well. As expected, coverage with one dose of MCV was constantly higher compared to coverage with two doses and thus susceptibility was inversely correlated. The long observation period may have also resulted in an underestimation of susceptibility in more recent birth cohorts due to very small numbers. Using coverage data from the electronic registry has resulted in a much better estimate of susceptibles of more recent birth cohorts. For the assessment of susceptibility with one or two doses of MCV we used the reported vaccine effectiveness for each dose as reported by the Centre for Diseases Control (CDC), Atlanta . The assAnother limitation of our seroprevalence data was that despite the entire region being covered, districts that are physically closer located to the laboratory were overrepresented and districts further away sent fewer samples. This is important as Tyrol is known to be a region with a high prevalence of vaccine hesitant persons . This maThe differences of susceptibility by birth cohort with seroprevalence and estimated adjusted susceptibility data from the electronic vaccination registry corresponded well in their pattern, and the mean varied between −5% for the second dose and 5% with the first dose. Susceptibility levels of seroprevalence data were higher in recent birth cohorts until 2009 compared to susceptibility calculated using coverage data. The latter was probably due to the small sample sizes of recent birth cohorts in the seroprevalence survey.The estimates of susceptible persons from the agent-based simulation model at a national level identified that approximately 70% of 19–30 year-old persons were protected with two doses of MCV as of 2018. In our data the increase of susceptibility using the second dose adjusted susceptibility started already from the birth cohort 2005 until birth cohort 1998 . This maLimitations of our coverage data were the different sources of numerator and denominator data. Participants who moved out of the region or died were still included in the numerator but were excluded from the denominator, in which only resident persons were reported. This might have led to an overestimation of vaccine coverage.Limitations of our surveillance data were that detailed information was only available since 2009. As older age groups were more affected in recent years in many countries, this may have resulted in underestimation of attack rates of younger individuals. If a virus is introduced to a susceptible population, cluster cases are known to occur in similar age groups. Therefore, the short observation period of our surveillance data may have reflected this phenomenon. Nevertheless, birth cohorts with more than one case reported were significantly positively associated with susceptibility levels > 10%.Targeted catch-up campaigns of the identified birth cohorts focusing on high incidence districts would definitely be crucial to close immunization gaps and reduce the number of susceptibles but raising awareness and social mobilization of the target groups were shown to be difficult. In our experience, when mandatory vaccination has not been implemented and no incentives for vaccination were provided, participation in catch-up campaigns remained unsatisfactory .We conclude the Tyrolean population is not sufficiently protected against measles and additional measures such as targeted supplementary immunization activities  are necessary to achieve elimination. This was also underlined by the fact that the majority of recently reported measles cases corresponded well with the birth cohorts with highest susceptibility levels. Priorities should be given to districts and birth cohorts with the lowest susceptibility, considering the WHO recommended values.Our seroprevalence data suggested that additional immunization activities should target birth cohorts between 2011 and 2015 as a first priority and birth cohorts between 1998 and 2010 as a second priority.Furthermore, our results suggest that birth cohorts before 1976 are not target groups for supplementary immunization activities as the majority of them are probably immune due to natural infection. In 2019, we conducted several activities mainly during the European Immunization Week to increase awareness and enhance uptake targeting the identified birth cohorts and districts, such as social mobilization and information campaigns in schools and universities, including health care workers.We conclude that using different sources of information such as vaccine coverage data, surveillance data, and retrospective seroprevalence data is useful to identify pockets of susceptibles and retrospective seroprevalence data could be an additional cheap and useful source of information to better identify sub-populations, such as birth cohorts and geographical areas with increased measles susceptibility. This will especially apply for older birth cohorts, whose coverage data are lacking.


Birth weight is a major predictor of infant growth and survival, and is dependent on maternal health and nutrition during pregnancy. This study aimed to determine the magnitude and identify nutritional and non-nutritional factors associated with LBW among newborn in Southern Ethiopia. Institutional-based cross-sectional study was used. Systematic random sampling was employed to select the study subjects. Data were entered into Epi-info Version 3.5.3 and then exported to SPSS Version 20 for analysis. Bivariable and multivariable logistic regression were used to compare birth weight across categories of independent variables. The output of the analysis were presented using adjusted odd ratio (AOR) with the corresponding 95% confidence interval (CI).A total of 358 mothers participated in the study. The mean (± standard deviation) birth weight of all term infants was 3304 (± 684) gram. The prevalence of LBW was 17.3% (95% CI 13.7–21.2%). Mothers who had MUAC less than 23 cm [AOR = 6.51 (95% CI 2.85–14.91)] and with hemoglobin < 11 mg/dl [AOR = 3.42 (95% CI 1.73–6.78)] have increased odds of delivering LBW and mothers who often take dairy products [AOR = 0.36 (95% CI 0.13–0.98)] were less likely deliver LBW babies than their counterparts. Birth weight is defined as the weight of the newborn, measured within the first hour of life, before significant postnatal weight loss occurs. Low birth weight (LBW) is defined by World Health Organization (WHO) as a birth weight less than 2500 g . AccordiA child’s weight at birth is the most important determinant of perinatal and infant mortality and morbidity and may have an influence on health in adult life. Based on international epidemiological observations infants weighing less than 2500 g are approximately 20 times more likely to die than heavier babies. Survived LBW babies are also more likely to suffer a high incidence of malnutrition, diarrhea, infection, neurodevelopment problems and physical defects. LBW babies are at high risk of developing chronic adult disease, such as type II diabetes, hypertension and cardiovascular disease later in their adulthood life .Ethiopian health and demography survey (EDHS) conducted in 2011 showed that among children born in the 5 years before the survey, 11% weighed less than 2.5 kilograms . StudiesAs birth weight determines the future health and health related factors of the newborn, intervening before and after pregnancy in appropriate maternal nutritional and non-nutritional factors is a window of opportunity for action.The study was conducted in Sawula town, Gamo Gofa zone, Southern Ethiopia. Sawula town is located 505 km away from the capital Addis Ababa and 285 km far from the regional capital Hawassa. The total population of the town is 43,639. The town has one district hospital, one comprehensive health center and four health posts. There are 5272 expected deliveries in respective health institutions which was calculated based on national conversion factor for estimated deliveries per year.Facility-based cross-sectional study was conducted from January to May 2016 in the health facilities of Sawula town. The study population was newborns of mothers who gave birth at Sawula district hospital and Sawula health center during the study period. Pregnant mothers with singleton live birth and aged greater than 15 years were included and mothers with preterm birth were excluded from the study.An independent sample size was calculated for both specific objectives. Accordingly, single population proportion formula was used to calculate sample for determining magnitude of LBW. In the calculation, 95% confidence level, expected proportion of 16.5% , 4% margInterviewer administered pretested questionnaire was used to collect the data. The questionnaire was prepared in English, then translated into the local language Amharic, and back translated to English in order to check its consistency.Maternal height was measured to the nearest 0.1 cm using ANC DETECTO® medical scale with sliding head metallic bar. The measurement was made without shoes, standing erect, feet together, heels, buttocks, and occiput touching the stadiometer and looking straight ahead. To proxy maternal nutritional status MUAC was used. It was measured to the nearest 0.1 cm using standard non-stretchable MUAC tape and taken on the middle left arm at relaxed position without any clothing and with optimal tape tension. Birth weight was measured using pretested and calibrated digital CROWN®baby weight scale as naked body. Birth weight was measured in the first 6 h right after birth and recorded to the nearest 100 g. All anthropometric measurements were taken in duplicates and the average of the two observations was ultimately registered.Venous blood from antecubital vein was taken for the determination of the hemoglobin and capillary blood for HIV test. Hemoglobin was determined by Sahli's haemoglobinometer (acid haematin method), and HIV test was done during labor by using HIV (1 + 2) Antibody Colloidal Gold (KHB) as a screening test, followed by HIV 1/2STAT-PAK® if positive. Where the result of STAT-PAK® is discordant with KHB, a third test, Unigold™ HIV is used as a tiebreaker to determine the result. For mothers whose HIV status known at ANC, card review was done to record HIV status of the mother on questionnaire.Data were entered into Epi-info Version 3.5.3 and then exported to SPSS Version 20 for analysis. The characteristics of the study subjects were analyzed using descriptive statistics. Frequency distribution were used to summarize qualitative variables. Numerical summary measures, mean with SD, were used to summarize quantitative variables after checking for the normality of data. The magnitude of LBW was estimated using proportion with 95% CI. Binary logistic regression was used to identify factors associated with LBW. Simple logistic regression analysis was used to identify candidate variables for multiple regression. At this, p-value < 0.25 was used as a rule of thumb for screening variables. Multiple regression was performed to determine the independent association between explanatory and outcome variable after controlling for the effect of confounding. The magnitude of association was measured using OR with 95% CI. The significance of association was declared at p-value < 0.05. Before declaring the association, the model fitness was checked using Hosmer and Lemeshow goodness-of-fit test statistic.Out of the total 380 sampled mothers who gave birth at two government owned health facilities in Sawula town, 358 participants responded to the questionnaire. This makes the response rate 94.4%. The mean (± SD) age of mothers is 26.9 (± 6.2) years and the highest proportion of respondents, 36.2% were between the age group 15–24 years. Most of mothers, 97.8% were married. About 75.5% were living in the family size of below five. Regarding place of residence, 56.4% of the respondents were living in rural area. Nearly half of the respondent mothers, 51.1% were illiterate and regarding their religion, 37.4% were protestants and 36.6% were Orthodox Christian  visit during the index pregnancy, of which 43.5% started the follow up in the first trimester. Nearly one third of deliveries were primipara.Among study participants, 14.0% and 2.2% had pregnancy induced hypertension (PIH), and gestational diabetes mellitus, respectively. Additionally, from 357 mothers screened for HIV status during pregnancy and labor, 1.7% were HIV positive. From 358 pregnant mothers screened for hemoglobin level at admission for labor 18.4% had Hgb level less than 11 mg/dl.Out of 358 postnatal mothers, 85.4% of them received nutrition education from health care providers during ANC visit they had in the index pregnancy. Of which 44.5% of mothers had practiced additional meal intake in the index pregnancy. Only 27.7% of mothers took iron tablets and 20.2% of mothers had deworming history in this pregnancy. Among participants, 45.3% of mothers were thin, MUAC less than 23 cm (Table The mean (± SD) birth weight of the all term infants was 3304 (± 684) g. The prevalence of LBW was 17.3% .After bivariable, multivariable logistic regression was considered to determine independent predictors on LBW. In multivariable analysis, hemoglobin level [AOR = 3.42 (95% CI 1.73–6.78)], consumption of diary product [AOR = 0.36 (95% CI 0.13–0.98)] and MUAC [AOR = 6.51 (95% CI 2.85–14.91)] had significant association (p < 0.05) with LBW (Table LBW is considered as a problem great enough to trigger public health action when its incidence exceeds 15% . In thisWe found that low hemoglobin level of mother during pregnancy was one of the predictor for LBW. Findings in Addis Ababa  and DebrOne of the risk factors for LBW in this study was low maternal MUAC. This finding was in consistent with a study in Sidama, Southern Ethiopia  and in KMaternal under nutrition due to an insufficient food supply has great effect on the growth and development of fetus. The finding in this study indicated that daily intake of dairy products at third trimester has positive influence on the birth weight of the newborn. Study in Sweden supported that low milk intake in the pregnant mother was associated with IUGR of the newborn . AnotherThe magnitude of LBW in this study is 17.3%. The significant predictors of LBW were maternal MUAC less than 23 cm, low hemoglobin level and frequency of daily intake of dairy products at third trimester. Strengthening quality nutrition education during ANC and improving the nutritional status of women particularly during pregnancy would reduce the occurrence of LBW. Therefore, health workers should provide routine antenatal iron tablet supplementation for as per the guideline, closely monitor maternal MUAC at antenatal visit and link cases for SFP services.The cross sectional nature of the data, which makes it impossible to draw inferences about the direction of relations among study variables. As the study was facility based; it is not possible to generalize the results to a particular population as compared to population-based studies. Recall errors and social desirability bias might have also occurred in the study.


The electric field in the hydrogen-bond network of the active site of ketosteroid isomerase (KSI) has been experimentally measured using vibrational Stark effect (VSE) spectroscopy, and utilized to study the electrostatic contribution to catalysis. A large gap was found in the electric field between the computational simulation based on the Amber force field and the experimental measurement. In this work, quantum mechanical (QM) calculations of the electric field were performed using an ab initio QM/MM molecular dynamics (MD) simulation and electrostatically embedded generalized molecular fractionation with conjugate caps (EE-GMFCC) method. Our results demonstrate that the QM-derived electric field based on the snapshots from QM/MM MD simulation could give quantitative agreement with the experiment. The accurate calculation of the electric field inside the protein requires both the rigorous sampling of configurations, and a QM description of the electrostatic field. Based on the direct QM calculation of the electric field, we theoretically confirmed that there is a linear correlation relationship between the activation free energy and the electric field in the active site of wild-type KSI and its mutants . Our study presents a computational protocol for the accurate simulation of the electric field in the active site of the protein, and provides a theoretical foundation that supports the link between electric fields and enzyme catalysis. Although it is still a controversial issue on the origin of the tremendous catalytic power of enzymes, it has been commonly accepted that the electrostatic field plays a key role in the enzyme’s high catalytic proficiency ,2. The ah is the Planck constant and c is the speed of light. The technique of VSE spectroscopy has been widely utilized to measure the electric field in proteins such as myoglobin ,9,10,11, follows ,18,19: simulations are now widely utilized in the study of proteins or other macromolecular systems to reveal more detailed information at the atomic level. Molecular mechanics (MM) force fields based on empirical potentials such as Amber, CHARMM, OPLS, etc. are still the predominant choices in MD simulations due to their inexpensive computational costs. However, despite the great success in applications of the classical force fields, they still have limitations in giving an accurate description of the electrostatic fields of H-bonds in proteins ,26,27,28Previous works ,26,27,28δ1 or Nε2 position, based on local electrostatic environment. The N-terminal and C-terminal were protonated and deprotonated, respectively. Force field parameters for 19-NT were obtained using the ANTECHAMBER [−1. The initial model of KSI bound with 19-NT used in this study was taken from the X-ray structure PDB id: 5KP4) . The mutKP4 . TheECHAMBER  module bECHAMBER  with HF/ECHAMBER . The proECHAMBER  algorithECHAMBER  method wECHAMBER  was appl19-NT probe of 19-NT were partitioned into the QM region  and 19-NT, where the contribution to electrostatic potentials from the probe C=O19-NT and the ring containing the C=O19-NT group was removed to avoid non-physical values, j is the atom number in residue i, j in residue i, 19-NT, and 19-NT bond is obtained as follows:19-NT, respectively, using Equation (2). 19-NT bond length.The saved structures from 40 ps QM/MM MD simulation  were used to calculate the electric field along the C=Outilized . The meaN denotes the number of residues in the protein, i and j represent the residue number.i and j (Gconcap) whose closest distance is less than a predefined threshold 19-NT and the ring containing the C=O19-NT group was removed) and solvent with the classical charge model is used to amend the QM results. The electric field exerted onto the C=O bond from 19-NT was computed using Equation (3), while the points where the electrostatic potentials were calculated are slightly off the atomic centers along the C=O chemical bond. Considering the expensive cost of the QM method, the single-snapshot approximation was adopted for the EE-GMFCC calculation of electric fields in KSI. Previous studies [The electrostatically embedded generalized molecular fractionation with conjugate caps (EE-GMFCC) method was employed to perform full QM calculations of electric field in KSI. The detailed description of the EE-GMFCC method can be found in our previous works ,53,54,55rk study ,46,55. T studies ,57 have For comparison, 2 ns classical MD simulations using the Amber ff99SB force field were also performed for WT KSI and three mutants , to calculate the electric field exerted onto the C=O group in 19-NT. In the classical MD simulation, the protein was placed in a periodic rectangular box of TIP3P water molecules, while the distance from the surface of the boxer to the closest atom of the solute was set to 12 Å. Counter ions were added to neutralize the system. The minimization and heating steps were the same as the equilibrium simulation for the previous QM/MM MD simulation. After that, a 2 ns MD simulation was performed to generate the trajectories for electric field calculations. The calculation of electric field was using Equations 2 and 3 with the charge model of the Amber ff99SB force field.The modeling of the H-bonds in biomolecules with an empirical method is very challenging because it involves important QM effects such as electronic polarization, charge transfer, and the many-body interaction. Many previous studies ,26,27,58To investigate why the conventional force field could not accurately describe the electric field in the H-bonding environment, we performed both QM/MM and classical MD simulations to calcualte the electric field for wild-type KSI and three mutants. The calculated electric fields based on classical MM simulations (Amber ff99SB) for all proteases are given in Previous works ,13,57,5919-NT probe and residues 16, 103 for WT and D103N are plotted in TYR1619-NT and OASP10319-NT in wild-type KSI, and OTYR1619-NT and NASN10319-NT in the D103N mutant. The hydrogen bond is supposed to be broken when the distance is greater than 3.5 Å. As shown in ASP10319-NT distance was mostly shorter than 3.5 Å, but the OTYR1619-NT distance was sometimes very large (greater than 5 Å). Furthermore, in the MM MD simulation of D103N, the OASN10319-NT distance sometimes also became very large (greater than 5 Å), while the OTYR1619-NT was mainly below 3.5 Å. However, the time scale of the MM simulation was much longer than that of QM/MM simulation. In the 40 ps time duration of the MM MD simulations for WT and D103N, the lengths of two hydrogen bonds could be both shorter than 3.5 Å, e.g., the first 40 ps of MM MD simulation of WT and the last 40 ps of MM MD simulation of D103N  of small model peptides, which does not include the electronic polarization and charge transfer effects for a specific chemical environment in folded proteins ,56. This19-NT probe experiences in the wild-type and three mutants  are calculated with the EE-GMFCC method and the empirical charge model of Amber ff99SB, respectively, based on the conformations sampled from the QM/MM MD simulations. The experimental work [19-NT VSE spectra, and increases of enzymatic unimolecular free energy barrier of these three mutants with reference to the WT, making them ideal candidates for theoretical investigation of the relationship between the electric field and the activation free energy. The correlation between the calculated electric fields using the EE-GMFCC method with single-snapshot approximation and the experimental values are plotted in R2 = 0.91. The computed electric fields with the EE-GMFCC method for the WT and the three mutants were consistently closer to the experimental data, as compared to those of Amber ff99SB. It also showed a good linear relationship between the computed electric field and the experimental observations with the correlation coefficient R2 = 0.96 for EE-GMFCC. Therefore, the trends in computed electric fields for WT and three mutants were both in good agreement with the experiment for EE-GMFCC and Amber ff99SB. It has been discussed in 19-NT experiences becomes larger when the QM effect is taken into account. When the electric field becomes larger, the C=O19-NT group would have stronger interactions with the hydrogen bonding residues, which explains why the QM/MM simulation could better maintain the H-bonds between residues 16, 103, and C=O19-NT than the classical MD simulation. It is essential to account for the QM effect in simulations of the electric field in the H-bond environment. In this section, the electric fields that the C=Otal work  by FriedThe predicted electric fields by EE-GMFCC are still smaller than the experiment for these four systems. There are two possible reasons for this: first, the protocol for the EE-GMFCC calculation did not include explicit conformational sampling at the QM level, due to the expensive computational cost, and only one plausibly representative structure was selected; second, the experimental electric fields estimated from the VSE spectra were based on the linear relationship of Equation (1), which may be overestimated by around 10% owing to neglecting higher-order terms. This has been demonstrated in the previous work based on the ab initio QM calculations of the vibrational frequency of the probe under the external electric field . Howeverkcat based on the transition state theory [R2) of 0.95 and 0.94, respectively. It has been suggested that the strong H-bonds between the residues 16 and 103, and the substrate at the active site of KSI have a significant impact on its catalytic rate [−), and thus possesses a larger dipole moment due to the charge rearrangement in the step of formation of the intermediate state  are tate see a,b. The 19-NT group forms two strong H-bonds with TYR16 and ASP103; in D103N mutant, one of the H-bonds previously formed between ASP103 and C=O19-NT in WT is replaced by the H-bond formed between ASN103 and C=O19-NT, in the Y16S mutant, the H-bond formed by TYR16 and C=O19-NT is removed as compared to WT, and in D103L mutant, the H-bond formed by ASP103 and C=O19-NT is removed. The structures of the H-bond networks for WT and three mutants are shown in TYR1619-NT are very similar in WT, D103N and D103L. Moreover, the distributions of the distance OASP10319-NT are also very close in the WT and the Y16S mutant. On the other hand, the distribution of H-bond length of OASN10319-NT in the D103N mutant has an apparent shift to a longer distance as compared to that of OASP10319-NT in WT, indicating that the H-bond becomes weaker due to mutation of Asp to Asn. These results show that the change of one H-bond has little impact on the H-bonding structure of another in the active site of KSI. The site-directed mutagenesis approach is widely utilized in theoretical and experimental studies on enzymes. It is worth investigating the structural influences on local chemical environment from the point mutation. In wild-type KSI, the C=O19-NT group, the electric fields exerted by other residues were all relatively small. The sum of the electric fields from residues 16, 103, and 40 (denoted as “Sum of 3RS” in 19-NT experiences were approximately the same in WT and the three mutants, showing that these three mutations had little impact on the solvent environment near the active site. The decomposed time-average electric fields of residues 16, 103, and 40 are listed in The ab initio QM/MM MD simulations were performed to calculate the electric field mainly exerted by H-bonds in the active site of KSI. By comparing with the simulations using the classical force field, we found that the H-bonds would be dynamically stabilized by the QM electronic polarization effect, which plays an important role in accurate calculations of the electric field. The electric fields of wild-type KSI and three variants were calculated using the EE-GMFCC method and the point charge model of the Amber ff99SB force field based on the configurations from QM/MM MD simulations, respectively. The calculated electric fields by EE-GMFCC were in good agreement with the experimental observations, while the results calculated with the Amber ff99SB force field are all much smaller than the experimental values. In spite of this, based on QM/MM sampling, both the EE-GMFCC method and the Amber ff99SB charge model could provide correct trends in the change of the computed electric fields for the wild type KSI and its three variants. By comparing the computed electric fields with the activation energies of these four proteins, a linear correlation relationship between these two physical properties is observed. The detailed structural analysis indicates that the point mutations in the three variants has minor impacts on the overall protein structure and the solvent environment. The change of electric fields in the three variants with respect to the wild type is mainly caused by the mutated residue itself. The electric field that plays a leading role in catalytic rate of KSI is mainly exerted by residues 16 and 103. This study presents a computational protocol for the accurate calculation of the electric field in the active site of the enzyme, and also provides a theoretical foundation supporting the link between electric fields and enzyme catalysis.