The Deferred Action for Childhood Arrivals (DACA) program provides temporary relief from deportation and work permits for previously undocumented immigrants who arrived as children. DACA faced direct threats under the Trump administration. There is select evidence of the short-term impacts of DACA on population health, including on birth outcomes, but limited understanding of the long-term impacts.We evaluated the association between DACA program and birth outcomes using California birth certificate data (2009–2018) and a difference-in-differences approach to compare post-DACA birth outcomes for likely DACA-eligible mothers to birth outcomes for demographically similar DACA-ineligible mothers. We also separately compared birth outcomes by DACA eligibility status in the first 3 years after DACA passage (2012–2015) and in the subsequent 3 years (2015–2018) - a period characterized by direct threats to the DACA program - as compared to outcomes in the years prior to DACA passage.In the 7 years after its passage, DACA was associated with a lower risk of small-for-gestational age  and greater birthweight  for births to Mexican-origin individuals that were billed to Medicaid. Estimates were consistent but of smaller magnitude for other subgroups. Associations between DACA and birth outcomes were attenuated to the null in the period that began with the announcement of the Trump U.S. Presidential campaign (2015-2018), although confidence intervals overlapped with estimates from the immediate post-DACA period.These findings suggest weak to modest initial benefits of DACA for select birthweight outcomes during the period immediately following DACA passage for Mexican-born individuals whose births were billed to Medicaid; any benefits were subsequently attenuated to the null. The benefits of DACA for population health may not have been sufficient to counteract the impacts of threats to the program's future and heightened immigration enforcement occurring in parallel over time.The online version contains supplementary material available at 10.1186/s12889-022-13846-x. On June 15, 2012, the Deferred Action for Childhood Arrivals (DACA) program was introduced by an executive branch memorandum . DACA waPrior research has identified the beneficial impacts of DACA on the health outcomes of recipients –7 and thNevertheless, there is evidence suggesting that the population health benefits of DACA may have been attenuated following direct threats to the program under the Trump Presidency and campaign , 15. HowBirth record data spanning 2009–2018 came from the California Department of Public Health’s Birth Statistical Master Files. Analyses were pre-registered at Evidence for Governance and Politics (EGAP) (20190605AB). The Committee for the Protection of Human Subjects, the institutional review board for the California Health and Human Services Agency, and Vital Statistics Advisory Committee approved the study protocol.We first restricted our data to approximately 3 years before DACA passage through approximately 7 years after DACA passage (June 2009–May 2018), Because there are no direct measures of DACA eligibility or recipient status in California birth records, we followed prior research , 17 and We restricted the sample to births for which vital statistics data indicated that maternal educational attainment was equal to or greater than high school completion or a GED by the time of delivery and maternal birthplace was one of the top 15 countries of origin for DACA recipients . As of 2We restricted the analytical sample to all live singleton births. We excluded birth records with gestational ages < 20 weeks and > 44 weeks and with birthweight for gestational age greater than 3 standard deviations from the sample mean . We exclSee eFigure 1 for the derivation of the analytic sample.> 37 weeks gestation), and binary outcomes of preterm birth , low birthweight , and small-for-gestational-age (SGA) [We evaluated continuous birthweight (in grams), and term birthweight  . We usedge (SGA) We used pregnant individuals’ birthdates to identify likely eligibility for DACA. This improves on the identification strategy of the prior national birth outcomes study that had maternal age rather than maternal birthdate, such that comparison groups were not closely centered around similar birthdates . Specifi> 5th birth). Models also included indicator variables for county, year, and month of birth.We controlled for maternal age and age-squared, a binary indicator of maternal educational attainment , infant sex, and parity  design, a quasi-experimental approach well-suited to examining the effects of policies among population subgroups while adjusting for secular trends in a “control” group of similar individuals . In partWe estimated linear models with robust standard errors in which we regressed each outcome on an indicator of whether the birth was to an individual who was likely DACA-eligible (vs. ineligible) based on their birthdate, an indicator of whether the birthdate fell in the pre-DACA period (June 2009 – May 2012) or the post-DACA period (June 2012 – May 2018), a multiplicative interaction term between these two indicators, and covariates. In order to shed light on potential differences in the long-run impacts of DACA, we alternatively tested a three-category indicator of whether the birthdate fell in the pre-DACA period (June 2009–May 2012), the immediate post-DACA period (June 2012–May 2015) or the period following the start of the Trump U.S. Presidential campaign (June 2015 – May 2018), The primary quantities of interest, which represent the association between DACA and each outcome, are the coefficients for the interaction term between the DACA eligibility and the birthdate timing indicator variables. Linear models for both continuous and binary outcomes are standard for DID analyses because of the different interpretation of interaction terms in non-linear models . CoefficWe evaluated results among all births regardless of payer type, and then among the subset of respondents with Medicaid as payer. Medi-Cal, California’s Medicaid program, covers prenatal care and labor and delivery for undocumented individuals. We therefore expected that those with Medicaid as payer were more likely than those with other insurance types to include both undocumented individuals eligible for DACA and their counterparts who held similar pre-DACA immigration status but were DACA-ineligible due to the arbitrary birthdate cut-off.Sensitivity analyses (summarized in eTable 1) were designed to evaluate central assumptions of the DID approach: 1) that trends in outcomes for treatment and control groups would otherwise be parallel if it were not for DACA passage and 2) that DACA passage did not contribute to changes in the composition of births for treatment or control groups in the 3 years post-DACA. We also evaluated whether observed associations could plausibly be driven by the one-year difference in average maternal age for DACA eligibility groups. We evaluated year-by-year changes in birth outcomes surrounding DACA passage by switching our binary pre/post-DACA indicator to an indicator of year of birth that spanned 2009–2018 but omitted 2012 given that DACA passage occurred during this year. We carried out the same difference-in-differences procedures as described above with the year of birth indicator. Finally, to shed light on potential mechanisms linking DACA and birth outcomes, we evaluated the association between DACA and prenatal care, using a measure of the number of prenatal visits reported on the birth record as our outcome. We elaborate on the details of these analyses in the Supplemental In the overall analytic sample, mean age at delivery was about 32 years; DACA-eligible individuals were an average of 1 year younger than their DACA-ineligible counterparts  for births to DACA-eligible individuals and 3322 g (SD: ± 520) for births to DACA-ineligible individuals. The mean length of gestation was 38.6 weeks (SD: ± 1.7).We found some evidence of association between DACA passage and birthweight outcomes in the years post vs. years pre-DACA, although these associations were largely concentrated among Mexican-born mothers and to those whose births billed to Medicaid Table .Table 2DAmong the overall sample, we found evidence of association between DACA and lower risk of small-for-gestational age . Among births to Mexican-born mothers, we found that DACA was associated with higher term birthweight  for DACA eligible individuals in the 7 years after the program’s passage.Among the subset whose births were billed to Medicaid, we found evidence of lower risk of small-for-gestational age  and greater term birthweight  for DACA eligible individuals in the years after DACA passage as compared to their non-eligible counterparts.Among births to Mexican-born mothers whose births were billed to Medicaid, we found evidence of lower risk of small-for-gestational age  and greater birthweight  and term birthweight  for DACA eligible individuals in the years after DACA passage as compared to their non-eligible counterparts.The results of models that instead used a three-category indicator to differentiate between the 3 years immediately following DACA passage and the years following the start of the Trump Presidential campaign . We found some evidence of association between DACA passage and the composition of births to DACA-eligible vs. ineligible mothers, although this evidence was the weakest for the subset of births covered by Medicaid, for which we observed the largest effect estimates (eTable 2). Specifically, in the overall sample there was evidence that DACA was associated with a lower probability of births being covered by Medicaid. We adjusted for this variable and all other covariates in our models to account for possible confounding.Associations between DACA and birthweight outcomes for births to Mexican-born mothers and billed to Medicaid were generally robust to “placebo” tests that a) moved the date of DACA implementation 1 year earlier, creating a false policy change date, b) utilized a false maternal birthdate cut-off  to designate mothers’ DACA birthdate eligibility, and c) switched the analytic sample to demographically similar U.S.-born individuals who should not have been impacted by DACA (eTable 4). Results were slightly attenuated but similar when we established a “wash-out” period that excluded births whose gestation spanned the pre- and post-DACA passage periods (eTable 5). We acknowledge, however, that in many cases confidence intervals estimated for placebo tests were overlapping with those estimated in our primary results.We re-estimated our primary analyses with a year of birth variable in place of the pre/post-DACA indicator (omitting the year 2012) . We focused this sensitivity analysis on the subset of births to Mexican-born individuals covered by Medicaid, given that the impacts of DACA appeared most salient for this group in our primary analyses.While year-by-year estimates were imprecise, these results suggest that there was some divergence in birthweight outcomes in the years post-DACA relative to the reference year of 2009. These differences appeared to be driven by declines in birthweight for those not eligible for DACA rather than by improvements in outcomes for those who were DACA eligible and were most apparent in the years 2014 and 2016, with attenuation of differences in 2017. We also note that, while pre-DACA trends were generally similar across DACA eligibility groups, there was evidence of divergent low birthweight outcomes in 2010 (as compared to 2009) between the two groups.Finally, analyses of the association between DACA and the number of prenatal visits showed no evidence of differences by DACA eligibility in the post-DACA period (eTable 7). Year-by-year analyses focused on Mexican-origin mothers covered by Medicaid  showed some divergence in the average number of prenatal visits, although not in the expected direction: those who were DACA eligible reported fewer prenatal visits than their DACA ineligible counterparts in the years following DACA, with significant differences in 2016 .This study provides some of the first evidence of the effect of DACA on birth outcomes and is the first to consider longer-term impacts of the program on birth outcomes during a period of direct threats to the DACA program. Our results suggest that DACA was associated with weak to modest improvements in birthweight outcomes among births to Mexican born individuals and those whose births were billed to Medicaid in the 3 years directly following DACA’s passage compared to 3 years prior. Findings of association between DACA passage and continuous overall and term birthweight were most consistent across analyses, although there was some evidence of association with lower risk of small-for-gestational age for the subset of births billed to Medicaid. These findings are notable, given that even modest differences in infant birthweight have been linked to a wide range of long-term health and developmental outcomes –25. NeveOur findings of stronger associations for the subgroup of DACA-eligible individuals with Medicaid as payer may have been driven by the fact that this group likely had a higher percentage of truly DACA-eligible individuals. This is because California has historically covered both prenatal care and labor and delivery for undocumented individuals under its emergency Medicaid program . Income-We found no evidence of association between DACA passage and the risk of preterm birth or low birthweight, and the parallel trends assumption appeared to be violated for analyses of preterm birth. The significant short-term associations observed between DACA passage for continuous birthweight outcomes and – in some analyses -- small-for-gestational age for births to individuals using Medicaid could suggest that any potential impacts of DACA may have operated through mechanisms specific to intrauterine growth restriction. Prior studies have suggested that associations between maternal economic and employment circumstances and birth outcomes related to fetal growth could be explained by biological mechanisms of impact on maternal immune and cardiovascular systems as well as on behavioral pathways such as smoking and physical activity during pregnancy , 28.We expected that increased access to care under DACA may be another mechanism of impact on birthweight outcomes. In particular, prior research has uncovered a positive link between Medicaid coverage and low-income women’s maternal and infant outcomes , 30. Immfewer average prenatal visits compared to their DACA ineligible counterparts. The reasons for this finding are unclear; future research might explore whether these results were driven by fewer pregnancy-related complications for DACA eligible individuals following DACA passage and/or whether these results differ in states without prenatal coverage for undocumented individuals under Medicaid.However, the results of our sensitivity analyses ran counter to our hypothesis, suggesting that the average number of prenatal visits following DACA passage were no different for those who were DACA ineligible as compared to those who were DACA eligible. Our year-by-year analyses suggested that in one of the years post-DACA (2016), those who were DACA eligible had Overall, our findings suggest weak to modest impacts of DACA on birthweight outcomes in the short run a specific sub-group. It is important to note that even small impacts on birthweight outcomes can have a wide range of beneficial implications for population health. Nevertheless, there may be multiple reasons for the fact that results were not consistent across birth outcomes and did not persist in the long-term. One potential reason could have been the countervailing adverse impacts of increased immigrant enforcement, including record-level deportations that took place in the U.S. in the years surrounding DACA passage . ResearcWe additionally found that differences in birthweight outcomes were attenuated to the null in the years following the announcement of the Trump U.S. Presidential campaign in 2015. Year-by-year estimates suggest that results for most outcomes may have been particularly attenuated in the year 2017, following the November 2016 election, although differences in the average number of prenatal visits persisted in 2017. These findings mirror those from a study that found that DACA was linked to improved self-rated health in the 3 years after the program’s passage, but that these benefits eroded after 2015 ; anotherThis study has several limitations, including the use of a proxy method for identifying DACA eligibility, which follows prior studies , 8, but Given the limited precision of our eligibility indicator, we were underpowered to pursue a more robust regression discontinuity design, which would have taken full advantage of the nature of the variation in DACA eligibility . HoweverIn addition, our analyses were limited to California. Although California is the U.S. state with the largest proportion of DACA recipients , resultsOn the other hand, although deportations were at an all-time high in the years surrounding DACA, California had generally refrained from passing anti-immigrant legislation in the years prior to DACA’s passage in contrast to many other states. Protective laws like DACA could have made less of a marginal improvement on population health in a setting already attempting to support immigrants regardless of legal status. A recent national-level study found associations between DACA passage and short-term improvements in birth outcomes . HoweverFinally, our analyses rest on a set of assumptions that we could not evaluate fully. Policy changes like DACA could have induced shifts in family planning and/or early pregnancy outcomes , leadingDuring a time in which DACA remains without Congressional protection, our study using California data suggests weak to modest short-term impacts of DACA on birthweight outcomes primarily for births to Mexican-born mothers that were billed to Medicaid. Associations with other birth outcomes were null. Any modest positive impacts on birthweight appear to have subsequently been attenuated during a period of heightened anti-immigrant rhetoric and direct threats to the future of the program. These findings have important implications for our understanding of the population health consequences of inclusive immigration policies given that even small improvements in birthweight outcomes may have important downstream consequences for health and development across the lifecourse. Nevertheless, the limited and short-term associations between DACA and birth outcomes identified in our analysis may reflect the fact that any substantial impacts of DACA on population health might have been attenuated by co-occurring restrictive immigration enforcement efforts and/or direct threats to the program’s future.Additional file 1.


C. difficile was quantified by qPCR; fecal metabolites and SIgA levels were measured by NMR and SIgA enzyme-linked immunosorbent assay, respectively. We assessed the putative causal relationships from birth events to gut microbiota and metabolites, and ultimately to SIgA, in statistical sequential mediation models, adjusted for maternal gravida status in 551 infants. As birth mode influences the ability to breastfeed, the statistical mediating role of breastfeeding status and milk metabolites was also evaluated. Relative to vaginal birth without maternal IAP, cesarean section (CS) after labor was associated with reduced infant gut SIgA levels at 3 months ; this association was sequentially mediated through gut microbiota and metabolites of microbial or milk origin. Mediating gut microbiota included Enterobacteriaceae, C. difficile, and Streptococcus. The milk or microbial metabolites in CS-SIgA mediating pathways were galactose, fucose, GABA, choline, lactate, pyruvate and 1,2-propanediol. This cohort study documented the impact of birth on infant gut mucosal SIgA. It is the first to characterize gut microbe-metabolite mediated pathways for early-life SIgA maturation, pathways that require experimental verification. How gut immunity in early life is shaped by birth in relation to delivery mode, intrapartum antibiotic prophylaxis (IAP) and labor remains undetermined. We aimed to address this gap with a study of secretory Immunoglobulin A (SIgA) in the infant gut that also tested SIgA-stimulating pathways mediated by gut microbiota and metabolites. Among 1017 Canadian full-term infants, gut microbiota of fecal samples collected at 3 and 12 months were profiled using 16S rRNA sequencing;  The most abundant antibody secreted by gut mucosa, secretory Immunoglobulin A (SIgA) represents a first-line immune defense against microbial infection ,2. The gBifidobacterium and Bacteroides, and raising risk of C. difficile colonization [Bifidobacterium quantity in infant gut microbiota [Early infancy is a critical period for the infant gut to be progressively colonized with resident microbiota that collectively drive gut maturation and immunity . Gut micnization . Intrapanization ,22. Whetcrobiota . Furthercrobiota ,25. Manycrobiota ,27. Thercrobiota ; microbicrobiota . Howeverwww.childstudy.ca (accessed on 1 April 2022)). Mothers of studied infants were enrolled in their second or third trimester of pregnancy from Vancouver, Edmonton, Winnipeg study sites between 1 January 2009, and 31 December 2012. They were retained in the study if their newborns were a singleton live birth at ≥35 weeks of gestation with a birth weight greater than 2500 g. In vitro fertilized births were excluded to avoid multiple gestation or preterm births, and home birth were also excluded due to the lack of documentation of maternal IAP. Hospital birth records provided data on infant sex, delivery mode, duration of the first and second stage of labor, maternal IAP and gravida status. Women colonized with vaginal Group B Streptococcus (GBS) or delivered by caesarean section were administrated by intrapartum penicillin or cefazolin, respectively. This information was used in this study to categorize infants into respective groups as follows: vaginal-no IAP , vaginal-IAP , CS-labor (labor phase occurred before CS delivery) and CS-no labor (scheduled CS delivery without labor phase). Information on maternal characteristics, such as ethnicity, age, depressive symptom, household pet exposure, education level, prenatal smoking and locations, as well as the infant characteristics, including gestational age, breastfeeding exclusivity and infant antibiotic use before 3 months and breastfeeding duration at 1 year were reported by standardized questionnaires completed by parents. Breastfeeding status at 3 months was categorized into exclusive-BF (breastmilk only), partial-BF (breastmilk supplemented with formula) or non-BF (no longer breastfed). Breastfeeding duration at 1 year was classified as <3 month-BF (ceased breastfeeding before 3 months), 3 to <12 month-BF (ceased breastfeeding between 3 and 12 months) and ≥12 month-BF (breastfeeding greater than 12 months). The demographic characteristics of the infants and their mothers were compared according to different birth events  birth cohort  and 12 months  were utilized to evaluate: gut microbiota composition, as characterized by diversity metrics and taxonomic abundance from Illumina MiSeq sequencing of V4 regional 16S ribosomal RNA (rRNA) processed by QIIME (v1.6.0) ; C. diff by qPCR ; metabol by qPCR ; and SIg by qPCR . SIgA dahttps://CRAN.R-project.org/package=vegan (accessed on 1 April 2022)). Differential abundance of microbial taxa by group was visualized by stack bar and tested using linear discriminant analysis (LDA) effect size (LEfSe) [p-value < 0.05, using vaginal-no IAP as the reference group for birth events, exclusive-BF and ≥12 month-BF as the reference group for 3-month and 12-month breastfeeding status, respectively. Using the same reference groups, C. difficile colonization rates by birth event were examined by Fisher’s exact test. Spearman’s rank correlations were performed between SIgA and gut microbial abundance following stratification by birth events and breastfeeding status. Microbial taxa found to be correlated with SIgA were further tested for their correlation with metabolite concentrations. Spearman correlation was conducted using the ‘corr.test’ function in R (https://cran.r-project.org/package=psych (accessed on 1 April 2022)). p values were corrected for multiple testing  using Benjamini–Hochberg adjustment [Statistical analyses were conducted using R  within RStudio . Maternal and infant characteristics were compared with Fisher’s exact tests. Linear regression modeling examined the association between birth events, or breastfeeding status and fecal SIgA levels. Values of fecal SIgA were Box-Cox transformed before linear regression to improve the normality of data [ (LEfSe) . Statistjustment , with R A schematic diagram of the sequential mediation model is shown in Potential covariates included pre-pregnancy depression, gravida status, pet exposure, infant age at stool collection and breastfeeding status when investigating the effect of birth events on infant fecal SIgA level ,37,38. AIn this large subsample of 1017 full-term infants from the CHILD general population birth cohort , there wp < 0.05, p < 0.0001, p < 0.0001, Infants born by CS with labor had a lower fecal SIgA level at 3 months compared to vaginal delivery without IAP exposure  were tested in mediation models , while statistically significant differences was not seen for CS without labor. This association with lower infant fecal SIgA levels was found to be sequentially mediated through breast milk constituents or metabolites produced by microbiota from breast milk constituents. The greatest mediating effect was seen with degree of breast milk feeding and infant fecal levels of galactose  or fucose . Both sugars are available to infants as free molecules from microbiota-metabolized human milk oligosaccharides (HMO) [In this general population of 1017 infants, we found medical intervention during birth to influence gut SIgA levels of young infants through pathways statistically mediated by microbes and metabolites. SIgA is a component of the intestinal microbiota-mucin barrier, critical to the exclusion of pathogenic microbes; it also binds to specific gut microbiota to stabilize its composition . Whereases (HMO) . They bees (HMO) ,44.E. coli and Bacteroides sp. from glutamate, which is plentiful in milk [Our study provides evidence for the role of milk-related metabolites other than milk IgA in also stimulating gut SIgA levels in infants since statistical mediation or the indirect effect via the metabolite path  was found to be separate from the direct effect of breast milk on SIgA levels . In this study, the direct effect was inferred to be SIgA supplied by breast milk, while indirect effects through microbes and metabolites were inferred to be pathways that stimulate endogenous SIgA production in infants. GABA was the other statistically significant milk-related mediator, attenuated by CS with labor. It is found in low levels in human milk but can be synthesized by lactobacilli,  in milk ,46,47. G in milk ,49. Gala in milk . C. difficile or Streptococcus (only lactate). These findings suggest that gut microbiota affect metabolite availability and in turn, SIgA production. Lactate is produced by bifidobacteria, streptococci and enterobacteria [C. difficile colonization following CS with labor or without labor lowered lactate and pyruvate, likely via metabolic consumption [C. difficile (more likely with formula-feeding) versus enterobacteria (more likely with breastfeeding) post cesarean delivery determines fecal levels of lactate and pyruvate, and ultimately SIgA production.Microbiota metabolites, lactate and pyruvate, statistically mediated the SIgA association with CS in the presence or absence of labor. They were in sequential mediation pathways that included infant gut Enterobacteriaceae, bacteria ; feedingbacteria . Pyruvatbacteria . Stool ebacteria , we intesumption , and subC. difficile, the latter of which lowered fecal choline levels, as also shown by others [Choline was another milk-related statistical mediator of the CS with or without labor-SIgA association. By far, it was the metabolite with the strongest direct association with fecal SIgA levels. Experimentally, choline deficiency leads to lowered fecal IgA levels . Choliney others . We founy others . In vitry others .Rothia of the Actinobacteria phylum were also identified as statistical mediators but of the path from infant exposure to maternal intrapartum antibiotics during vaginal birth to higher fecal SIgA levels. Unique from other mediating microbiota, Rothia are also found in oral microbiota, more commonly during formula-feeding [Rothia possess enzymes to break down complex dietary proteins such as gluten [Rothia are found to be enriched in both oral and fecal microbiota of patients with COVID-19 infection [Rothia metabolites, lactate and to a lesser extent, 1,2-propanediol [Rothia pathway to SIgA induction.Genus -feeding . Their f-feeding ,70. Roths gluten . It is infection . In our panediol ,74, wereThe CHILD Cohort Study, possessing a large, comprehensive and representative dataset, facilitated the thorough assessment of gut microbe-metabolite mediated pathways from birth mode to gut SIgA. We pioneered exploration of statistical sequential mediation between birth mode and SIgA levels in early life, the window of opportunity for gut microbiota modulation of immune development. The sequential mediation model allows mediators to be linked serially in a causal way rather than in parallel, which enabled testing of putative causal relationships between independent, multiple mediating and dependent variables. The large dataset included adequate sample size to ensure model adjustment for early life covariates in assessments of 3-month SIgA and metabolites. Sample size was not sufficient for assessment of SIgA at 12 months. Further, we lacked quantification data on breast milk SIgA, and thus, cannot exclude the influence of SIgA in mother’s milk. Our 16S rRNA gene sequencing approach also limited taxonomic resolution at the species and strain level. Finally, the biological plausibility of the reported mediating pathways requires experimental verification.Through statistical mediation methods we identified candidate metabolic pathways involving galactose, fucose, GABA, choline, lactate, pyruvate and 1,2-propanediol to altered gut immunity in young infants following cesarean section delivery. When breastfeeding is less likely following cesarean, the reduced availability of these milk and microbiota metabolites leads to lowered SIgA levels in the infant gut. Hence, the benefit of breastfeeding on gut immunity following cesarean delivery is not only direct provision of SIgA to the nursing infant but also promotion of an infant’s own gut SIgA production through milk-related metabolites. Raising levels of metabolites, such as choline, has many additional benefits, including protection against impaired infant brain development in pregnancies complicated by viral infection .


This study aimed to explore the effect of body mass index (BMI) on neonatal outcomes in patients with polycystic ovary syndrome following the frozen embryo transfer (FET).2), normal weight (BMI between 18.5 and 24.9 kg/m2), and overweight (BMI between 25.0 and 29.9 kg/m2). Logistic regression models with generalized estimating equations were used for clustering by patients to explore the effect of BMI on neonatal outcomes.This study included 1,676 singletons born from mothers with polycystic ovary syndrome (PCOS) after FET between 1 Jan 2007 and 31 Dec 2019. BMI was categorized into three groups: underweight  babies  significantly decreased for underweight mothers and significantly increased  for overweight mothers. The rate of high birth weight among infants from overweight mothers  was significantly higher than those from normal-weight mothers after adjusting for known confounding factors. The rate of small for gestational age (SGA) singleton  was lower among underweight mothers than normal-weight mothers.Maternal underweight was a significant protective factor against LGA infants for singletons born from patients with PCOS after FET, whereas maternal overweight was an adverse factor for LGA infants. Polycystic ovary syndrome (PCOS) is one of the most common endocrine disorders affecting about 5–20% of reproductive-aged women . It is dEvidence shows that PCOS has a negative impact on neonatal outcomes. Infants born from mothers with PCOS are at increased risk of preterm birth, large for gestational age birth (LGA), admission to a neonatal intensive care unit, or perinatal mortality –10. Resein vitro fertilization cycles reported that those with FET had a higher live birth rate and a lower risk of ovarian hyperstimulation syndrome than those with fresh embryo transfer  is being used as a great substitute for fresh embryo transfer in clinical treatment. A multicenter randomized controlled trial among 1,508 infertile women with PCOS undergoing the first transfer . HoweverTherefore, this retrospective cohort study was performed to evaluate neonatal outcomes among underweight and overweight vs. normal-weight patients with PCOS following the FET.All data used in this retrospective study was extracted from the clinical database at the Assisted Reproduction center of the Shanghai Ninth People's Hospital . Cycle-level information on patient characteristics, details of clinical treatment with ART procedure, and pregnancy and neonatal outcomes were recorded in this database . A serie2) and was divided into three categories according to the classification of the World Health Organization: underweight (BMI less than 18.5 kg/m2), normal weight (BMI between 18.5 and 24.9 kg/m2), and overweight (BMI between 25.0 and 29.9 kg/m2). In our center, the number of patients having a pre-pregnancy BMI of >30 kg/m2 was small; hence, we did not include these patients in this study. The following reasons can account for the small number of obese women. On the one hand, the BMI of the Asian population is generally lower than that of non-Asian populations /(height (m)ulations . We usedth and 90th percentiles for gestational age.For the neonatal outcome, clinical data of gestational age and birth weight were collected. The gestational age for cleavage-stage embryo transfer was calculated by pulsing 17 days since the date of embryo transfer, while the gestational age for blastocyst transfer was calculated by pulsing 19 days since the date of embryo transfer. Preterm birth was defined as the delivery before 37 completed gestational weeks by the World Health Organization, and gestational age of lesser than 32 weeks was categorized as very preterm birth. The birth weights of <2,500 and >4,500 g were categorized into the low birth weight group and the high birth weight group, respectively. LGA and SGA were defined as birth weight over the 90th percentile and below the 10th percentile for gestational age, respectively, according to the Chinese sex- and gestational age-specific birth weight standards . ApproprThe distributions of basic characteristics in different BMI categories were computed. Then, the neonatal outcomes for women with different BMI categories were contrasted. Characteristics of the cohort and neonatal outcomes were described using mean with standard deviation (SD) for continuous variables and number of cases with percentages for categorical variables. The between-group differences were tested using the one-way ANOVA test for continuous variables and the Pearson chi-square test for categorical variables. Then, we used logistic regression models with generalized estimating equations for clustering by patients to explore the effects of BMI on neonatal outcomes. The unadjusted and adjusted odds ratios (OR) and corresponding 95% confidence intervals (CIs) were reported. Maternal age, infertility type (primary or secondary infertility), infertility duration, parity (no child/one/two or more children), ART procedure type , embryo stage at transfer (day 3 or day 5/6), number of embryos transferred (1 or 2), infant gender (boy or girl), and year of birth were covariates adjusting for analyses. All statistical analyses were performed using the statistical package Stata, version 12  and a two-sided 5% level of significance.A total of 1,676 live singleton births from mothers with PCOS after FET were enrolled in the study. Of these singletons, 101(6.02%) were delivered by mothers with underweight, 1,148 (68.50%) by mothers with normal weight, and 427(25.48%) by mothers with overweight. The basic characteristics of mothers across BMI categories are shown in The neonatal outcomes according to maternal pre-pregnancy BMI are shown in The logistic regression analyses were used to explore the effect of pre-pregnancy BMI on neonatal outcomes . The risThis study focused on the effect of pre-pregnancy BMI on neonatal outcomes for singletons from mothers with PCOS following FET, and the result showed that maternal underweight was a significant protective factor against LGA infants for singletons born from patients with PCOS after FET, whereas maternal overweight was an adverse factor for LGA infant. In addition, the risk of high birth weight was significantly higher for infants from overweight mothers, and the risk of SGA babies was higher for infants from underweight mothers, compared with infants from normal-weight mothers.Excessive fetal growth can result in LGA infant or fetal macrosomia, which substantially raises the risk of short and long-term morbidities and mortality –24. In tOur findings suggested that pre-pregnancy underweight increased the risk of SGA infants. A similar result was reported by Zhang et al.  explorinPreterm birth is related to a high risk of neonatal mortality and morbidity . SurvivoThe main strength of this study was that we concentrated exclusively on singletons born from women with PCOS conceived by FET. The finding of this study offered supporting evidence for evaluating the effect of pre-pregnancy BMI on neonatal outcomes. The dataset obtained from the single center avoided the potential interference from different ART techniques and laboratory environments in centers on neonatal outcomes. However, the study has several limitations. The first inherent limitation of this study was the retrospective design; however, strict inclusion criteria have been set to ensure high-quality data. The second limitation was that we did not explore the association between obesity and adverse neonatal health in this study owing to the small number of obese women with PCOS. The third limitation was that some potential confounders could not be adjusted including maternal nutrition conditions, educational status, and the type of PCOS. Finally, no long-term follow-up was conducted among offspring from mothers with PCOS, thus great attention should be devoted to the growth, development, and psychosocial health of offspring in the future.In conclusion, this study provided scientific evidence that maternal pre-pregnancy BMI affected the birth weight of singletons born from mothers with PCOS after FET cycles. For women with PCOS getting pregnant by FET, being overweight was a risk factor for having an LGA baby, whereas being underweight was a protective factor for having an LGA baby. In addition, being overweight was significantly associated with an increased risk of high birth weight, and being underweight was related to an increased risk of SGA. These findings are vital for clinical practice. Clinicians should encourage women with PCOS to optimize their BMI before fertility treatment by taking some measures including dietary adjustment, physical exercise, and behavioral intervention. Maternity and newborn care providers should pay more attention to women with PCOS with abnormal BMI for the prevention of adverse infant outcomes.The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.The studies involving human participants were reviewed and approved by the Ethics Committee  of Shanghai Ninth People's Hospital Affiliated to Jiao Tong University School of Medicine. The patients/participants provided their written informed consent to participate in this study.QZ and HGu designed the study and drafted the manuscript. HGa and BW contributed to data collection and analysis. QZ revised the article. All authors read and approved the final manuscript.This study was funded by the National Natural Science Foundation of China  and the interdisciplinary program of Shanghai Jiao Tong University (YG2019QNA19).The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


The paper presents a new analytical four-layer (air–water–bottom–non-conductive layer) horizontal electric dipole model which allows an accurate approximation of ship's Underwater Electric Potential (UEP) from a sufficient depth in shallow coastal marine waters. The numerical methods, usually Finite Element Method (FEM) or Boundary Elements Method (BEM), are typically used to estimate the electric field and the distribution of static electric components of UEP around the ship. These methods enable analyses with high accuracy but, compared to other point-electrode methods and the proposed analytical model, they are relatively complex and need high computational time. The developed analytical model proposed in this paper allows real-time calculations without significant loss of accuracy of the UEP estimations. In the model, the problem of boundary values at the borders of individual layers is solved using the reflection/image method and applying the idea of continuity of electric potential at a given boundary between two adjacent layers. Its accuracy is verified based on the synthetic data provided by specialised software packages making use of FEM and BEM numerical methods. A dimensionless quantitative analysis of the relationships between basic parameters of the proposed four-layer analytical model and their impact on the accuracy of representation of individual electric field strength components is also delivered. The relationships between water and bottom conductivity and between water depth and bottom thickness are investigated and described. The obtained results show that the developed model allows detailed and reliable analysis of the electric field, especially in shallow coastal waters. In such applications, currents with a sinusoidal or rectangular waveform are usually applied to the analysed object. The resulting voltages are measured using the surface electrodes to assess the interior conductivity distribution and the permittivity distribution. Not only man-made technical devices use electric field technologies. For example, in the natural world, similar techniques are used by weakly electric fish. For food acquisition and navigation purposes, those fishes can generate an electric field and use echo analysis with receptors placed on their bodies. This natural phenomenon, called biological electrosense, is extensively studied by numerous researchers. For example, mathematical models for weakly electric fish in terms of active electrolocation can be found in5, while those concerning shape recognition and classification in6. Advanced optimization or discontinuity testing methods are used in science to find the structure and parameters of such models. However, it is hard to expect living organisms to act in this way. The papers7 and8 present a simplified approach and mathematical framework in the form of first-order polarization tensor for the purpose of object electrosensing, while9 shows the implementation of an algorithm for estimating the size, shape, orientation, and location of an ellipsoidal object at short range in the robotic active electrosense system.For years, electric field related technologies have been of interest in various fields of science and technology such as geophysics, archaeology, industrial and chemical process monitoring, and biomedical engineering. Meaningful applications of these technologies may be found in systems used in Electrical Impedance Tomography (EIT)15, in predicting UEP signatures to evaluate a possible risk of the vessel beaning detected by naval mines16, in geophysical surveys to obtain the information on seabed structure18, in the exploration of oil and gas deposits located in the seafloor20 in the tasks of ships tracking21 and object localization below the waterline22, or in ships classification23.In this article, the authors focus on applying underwater electrical potential techniques in marine industry. The measurement of underwater electrical potential (UEP) is widely used in civil and military marine applications for different purposes. For example, it is used for monitoring a ship hull corrosion and its protection against corrosion24. Along with magnetic anomalies and hydrodynamic variations detected in the presence of the ship, the static electric component of the electric field is one of physical fields referred to as ship signatures26. The authors have experience in analyzing and modeling the magnetic field distribution with the use of a multi-dipole model for both synthetic data28 and real data from measurements29.The electric signature of maritime platforms is a measure of the static electric field generated by the electric currents flow around the vessel's hull. These currents are generated by galvanic corrosion of the hull  or, as anti-corrosion currents, by Impressed Current Cathodic Protection (ICCP) systems or passive protection systems based on the usage of sacrificial anodes, designed to prevent corrosion (cathodic ship protection state). The currents flow between particular ship's hull components made of different electrochemical materials when they are immersed in seawater. In this case, the seawater acts as an electrolyte, and the electric circuit is closed through the metallic hull of the ships or even through its internal cabling system. The source electrode (anode) in this circuit is the submerged part of the ship's hull and sacrificial anodes of the passive cathodic systems, while the sink electrodes (cathode) are the shaft propellers and hull damages. From a sufficient depth, this phenomenon can be approximated by a single horizontal electric current dipole system25. Hence, to decrease the ship’s electric field, the process of optimizing its structure is used, along with the impressed current compensation30 and advanced ICCP systems11. The authors of the paper31 compared those methods and showed that the impressed current compensation method should be used to achieve the maximum reduction of the ship’s static electric field signal.The ship's UEP signatures depend on the type and state of the used materials, their locations along the ship and the size of the galvanic components installed on the ship's hull, which constitute the object-specific configuration of anodes and cathodes. For a ship equipped with cathodic protection, its UEP signature is much more significant  and complex (number of peaks) due to the presence of additional anodes and currents generated by external power systems. The UEP signatures are of particular importance for the safety of ships triggering sea mines or other vessel detection systems  equipped with integrated electric field sensors, which are characterised by high sensitivity and low measurement drift and enable UEP monitoring over a wide frequency rangeσ2 and σ3, respectively). Until now, numerous authors have already described two-layer and three-layer models of the horizontal current dipole but in those cases they had to adapt appropirelly the models parametres to handle the bottomsea characteristics32. While, according to the best of authors' knowledge, the four-layer model has not yet been presented in the literature.This paper presents and verifies, using synthetic data as the reference, a new analytical four-layer (air–water–bottom–non-conductive layer) horizontal electric dipole model—Fig. 35. Additionally, when considering the fourth non-conductive layer with a particular thickness below the bottom, characteristic flattening of the electric field at layer’s boundary without deep penetration into the bottom layer is observed—Fig. 36. Depending on sea currents, this layer of sediments can vary in thickness, and in addition, depending on the dominant material, it can have different electrical conductivity, which is strongly related to the distribution of the electric field generated by the ship. Furthermore, the mixing of seawater with freshwater, e.g. in river deltas, affects the chemistry of the seawater, which can additionally influence the potential recorded by electric sensor electrodes35.In shallow water and nearby seabed, due to conductivity difference between water and bottom, a specific reflection of the electric field takes place on the boundary between these two media37, SIMSEN38 from SAES39, BEASY Corrosion40, or COMSOL Multiphysics41, which make use of Finite Element Method (FEM) or Boundary Elements Method (BEM) methods, are typically used to model the electric field distribution and the UEP signature around the ship. These methods enable analyses with high accuracy42 and can be used to estimate the electric filed of a maritime platform from the definition phase up to the sea trials. However, compared to other point-electrode methods and the model presented in this paper, they are relatively complex and need high computational time42. At the same time, the proposed four-layer model enables real-time calculations without significant deterioration the accuracy of the obtained results. In this model, the problem of boundary values at the borders between individual layers is solved using the "reflection/image" method and applying the idea of continuity of electric potential at a given boundary between two adjacent layers.Specialised software packages, such as OPERA32, the author has described the electromagnetic field generated by a horizontal electric dipole in the three-layer region. The complex equations field formulas included all points in all three layers. The model of electric dipole and the inversion method were used to get an equivalent ship's electric field. The authors of the paper42 have presented the ships' UEP model based on the multiple point-electrode method which allowed reconstructing more precisely ship’s electric signature close to the hull than the single point-electrode (single dipole) method. They used the Particle Swarm Optimization (PSO) method to solve the number, positions, and current values of point-electrodes according to the UEP distribution at a known depth below the keel of a ship. Based on the scaled ship model experiment, the authors have shown that the resulting model can be used for simulation and prediction of UEP signatures.In37, SAES39) making use of FEM and BEM numerical methods, respectively, to model a single horizontal current dipole in the marine environment with shallow water , may be computed by taking into account the gradient of the potential V in appropriate direction  relative to the center of the Cartesian coordinate system, h represents the depth of their placement, and 2a denotes the distance between the electrode with current I (positive) and that with current − I (negative).In order to derive the analytical model of the electric current dipole and calculate its underwater electric potential, let us consider two spherical electrodes with current ace Fig. a. In FigVI and individual components Ex, Ey and Ez of the electric field intensity only in the water layer  should be introduced to solve the electric field in those two areas. The virtual electrode I′ is introduced in area II  allows to calculate the electric field in the space which is only valid in area II with electric conductivity σb  in layer 1—air , a pair of virtual electrodes I′ has to be introduced to satisfy boundary conditions BC2 (z = − 2hw underwater − second reflection), but on the other hand these electrodes disturb the boundary condition BC1. Therefore, in the next (third) reflection, a pair of virtual electrodes I′ should be introduced at height 2hw (z =  + 2hw above water) Fig. hw (z = − 4hw underwater) and the next mirrors ± 2nhw (n—number of reflections) should be taken into consideration.Now let us consider in detail the four-layer model Fig. . To takeair Fig. a. In theI and I′ affect layer 2 (water)—as they generate an electric field intensity only in this layer, which is presented in Fig. The electrodes with currents I″ has to be introduced to satisfy boundary conditions BC3  (first reflection) to satisfy the boundary condition BC3, and at the second reflection point z =  + 2(hw + hb) (above water) to meet the boundary condition BC1  (i—number of reflections) should be taken into consideration to satisfy boundary conditions BC1 and BC3 and to ensure the required accuracy of calculations. Note that the values of particular currents do not change and do not depend on the number of reflections; only the values of successive reflections increase for an increasing distance of the virtual electrodes from the electric field sensor located in water.In the case of four layers (air–water–seabottom–nonconducting), the pair of virtual electrodes BC3 Fig. c. To takI, I′ and I″, Eqs. I, I′ andEx, Ey, Ez caused by current I and virtual currents I′ and I″ for the two considered electrodes , then the currents I′ = 0 and I″ = I, and the boundary condition BC3 is met. If, e.g. the bottom electric conductivity σb and the water conductivity σw are different but hb tends to zero, the four-layer model is also reduced to the three-layer model . The expressions in parentheses in formulas  an anσb is formulas , and theI (both for positive and negative electrodes) exist in the model, without any additional virtual electrodes and currents. In this case, only three Eqs. (n = 0) and are finally considered in the two-layer model. Hence, only four source currents generate the electric field in water , and the depth of electrodes h is 1 m. In this case the use of additional mirror reflections (n ≥ 1) is unnecessary.On the other hand, if the seabed is very deep, the four-layer model is reduced to the two-layer model (air and water layers—without bottom and non-conductive layers), then only two electrodes with current ree Eqs. , 24) an anI , the number of reflections should be suitably increased to ensure the required accuracy of calculations.The number of additional mirror reflections becomes crucial when electric sensors are near the bottom with lower electrical conductivity then that of seawater: n = 10 reflections was assumed, and then the appropriate values of dimensionless components of electric field intensity Ex  and Ez  for n = 1, …, 10 were calculated as:In these tests, the reference number of n for water depth hw = 10 m, and bottom depths hb = 1 m and 0.1 m. Additionally, it was assumed that the electric conductivity of water σw and bottom σb was equal to 4 S/m and 0.4 S/m, respectively, the electrodes with source current equal to 1 A were placed at depth h = 1 m, and the distance between the electrodes 2a was equal to 1 m. Notice that in this experiment, the electrodes were placed symmetrically to the center of the considered Cartesian coordinate system (x-axis), and the y-coordinates of both electrodes were equal to 0 (y = 0). That is why only the dimensionless components of electric field intensity , while the minimal and maximal values of component Ez are equal to each other.The values of dimensionless components of the electric field intensity Eqs. –32 are pity Eqs. –32 were n = 0), then the errors in electric field intensity calculations can reach even 100% . The area of 100 m × 100 m × 50 m was taken as the test field, and the tangential electric field boundary condition was assumed at the field boundaries. The air area was not modelled in the FEM models. The area along the z-axis was divided into twelve sub-areas (sub-layers) with various thicknesses to carry out successive numerical experiments efficiently. From top to bottom, the first layer was 5 m thick, the second was 4 m thick, and the subsequent six layers were 1 m thick each. The ninth to tenth layers were 5 m thick each, and the final twelfth layer was 20 m thick. The conductivity parameters of those sub-areas (sub-layers) were changed in such a way as to obtain finally different thicknesses of the water layer, the bottom layer, and the non-conductive layer in the analyzed numerical model, without introducing additional changes in the numerical model mesh. The model built in the Opera environment is shown in Fig. xz surface.The FEM models were built in the Opera simulation software39. In the BEM method, the air and water areas do not need to be modelled, and therefore the entire area is modelled as an empty box, thus reducing the number of elements. In BEM, the length of the limits is conditioned by the length of sources and the value of estimated electric field. For this study, the spatial limits of seawater were modelled using a box with dimensions of 100 × 100 m and the height equal to water depth (variable parameter in subsequent experiments). The air–water boundary was modelled by means of a horizontal plane of symmetry defined at depth equal to 0 m, boundary between water and air. The contour of the seawater area, close to the bottom, was meshed using 6960 elements with a different grading, increased the number of elements in the area close to the position of the source and measurement points. The horizontal plane of symmetry was not meshed. The monopoles were modelled as point sources with positive (anodes) or negative (cathodes) currents. All these elements and sources were defined inside of only one zone for the two-layer models (air and water) in the BEM model. When the bottom conductivity and its thickness were taken into account in the analysis of the three-layer model , another zone had to be defined for the bottom in the model already including two zones. For this analysis, the bottom was modelled as a box with the same dimensions than the water model and the height equal to bottom thickness, and meshed with 200 equally spaced elements. The boundary condition defined for both models was zero current density and voltage at one element of the end of the box. Figure The BEM models were built in the simulation software developed by SAES44) proposed in “It is essential to point out that each change in scenario characteristics  of the FEM or BEM models leads to new numerical models which have to be built in each software environment and solved again to obtain new results for the electric field components. All these operations take incomparably much more time than the calculations making use of the four-layer analytical model . The parameters of the analytical model used in this comparison analysis are shown in Table y = 0 m and y = 15 m. For y = 0 m, the component Ey of the electric field intensity is equal to zero due to system symmetry and therefore is not presented in separate figures, nor is it noted in Table x =  ± 100 m . The errors on the border of the SAES numerical model (BEM) shown in Fig. The distributions of three components of the electric field intensity Figs.  and 14 aThe next subsection analyses the electric field intensity results regarding the electric conductivity of the water and bottom, the depth of the water, and the thickness of the bottom. This analysis was carried out using the full analytical four-layer model.The second set of tests increases the complexity of the analysis. In these tests, the water depth and the bottom depth were assumed finite Fig. . The parx =  ± 100 m) for the SAES model (BEM)—except the case for Ez and y = 15 m presented in Fig. The distributions of three components of the electric field intensity Figs. , 23, 24 hb = 1 m and hb → ∝) on electric field  is presented in Figs. The previous section “” has shoIn this section, further comprehensive analyses are presented to show model’s ability to reproduce the electric field intensity in shallow and deep seawater conditions with different seabottom structure and parameters. The main objective of these analyses was to verify how the components of the electric field intensity in seawater depend on the relations between electric conductivities of water and bottom , th, th40), Ex, Ey and Ez as waveforms covering twelve cases  and three different values of dimensionless yu ∈ ) according to the changes of dimensionless bottom thickness huw. In contrast to the test results presented in “Ey was also included in the analysis, since in this section, the electric field intensity components were also analysed for different values of yu .The third analysis concerned the case of 4 layers: air–water–bottom–non-conductive layer, with constant thickness of the bottom layer. The obtained results are summarised in Figs. huw greater than 2.5 . Therefore, it can be concluded that the electric conductivity of the bottom is essential, especially in shallow sea waters  >).It can be seen that for the dimensionless water depth .5 Figs.  and 35, Ex, Ey and Ez as waveforms covering twelve cases  and three different values of dimensionless yu ∈ ) according to the changes of dimensionless bottom thickness hub.The fourth case of the performed analysis concerned the 4-layer model (air–water–bottom–non-conductive layer) with constant thickness of the water layer. In this set of tests, the dimensionless water depth was assumed equal to 1. The obtained results are summarised in Figs. Ex is larger by about 60%, and the component Ez is smaler by about 50% than the corresponding components in the ‘infinite’ water case (Figs. Ex decreases and the component Ez increases when the bottom thickness increases and the electrical conductivity of the bottom takes larger values. Generally, the increase of Ez (Fig. Ex (Figs. Ey (Fig. hb is more significant for higher values of bottom electric conductivity σw. The differences between the values of electric field intensity components reach even as much as 60% (Fig. hb is essential and has to be taken into account, especially when this thickness is relatively small and the value of bottom conductivity of σb is close to that of water conductivity σw.It can be seen that for the dimensionless botom depth equal to 0, the electric field intensity component se Figs.  and 39. Ex Figs. , 37 and  Ey Fig.  as a res60% Fig. . It can In the technical literature, the electric field intensity is usually described using a three-layer model of horizontal electric current dipole. In the model proposed in this paper, a fourth layer is introduced to allow a more detailed and reliable analysis of the electric field, especially in shallow coastal waters. The fourth layer has special interest when the sea bottom has an estimated thickness, and it is needed to be introduce this environmental characteristic in the model. Due to the attenuation law of electromagnetic waves in seawater, the electric field decreases dramatically with the distance, therefore, the electric signatures are considered as near field signatures. Hence, the electric field propagation has especial interest in shallow coastal waters areas where the distance between the source and measurement system are limited by the water depth.The performed analysis has shown that the proposed four-layer analytical model of horizontal electric dipole can be used for fast and reliable calculations of the electric field in seawater in the presence of the fourth layer. Furthermore, the performed simulations tests and their results have shown that the bottom thickness and its electric conductivity significantly affect to the electric field's intensity distribution in shallow seawater. The comparative analysis of the results obtained from the proposed analytical model and from FEM and BEM numerical models confirmed the correctness and accuracy of the four-layer model. The differences of results between the numerical models and the analytical model were only at the level of a few percent. The FEM and BEM numerical methods have been designed to be used from the definition phase up to the sea trials. These models can use polarization curves or point sources and due to the amount of mesh points, require large computational time to solve electric field models. For this reason, these methods cannot be used for real-time operations, being the analytical models widely used. The analytical four-layer model performs the estimations of UEP faster than the numerical methods, almost a thousand times and therefore can be used for real time operations solving models based on dipoles.Future works will focus on the use of the presented analytical model of horizontal electrical current dipole to develop a multi-dipole model of ship's electric field which will allow reconstructing its electric signature in any maritime area. In that case, the ship's electric signature should be measured on a measuring range with known values of seawater depth, bottom thickness, and electric conductivity of seawater and bottom. Once its electric signature is known from those experiments and its multi-dipole model parameters are identified, the electric signature of the ship may be reconstructed for any maritime conditions. This kind of procedure may be used to protect ships against sea mines, or other underwater detection systems triggered by making use of electric field intensity measurements. This problem is relevant not only for naval ships but also for civilian transport vessels operating in shallow coastal waters.


To compare efficacy and safety of the levonorgestrel-releasing intrauterine system (LNG-IUS) with systemic medication or blank control in the treatment of dysmenorrhea.PubMed, EMBASE, the China National Knowledge Infrastructure (CNKI) and Wanfang Data were searched to collect randomized controlled trials (RCTs) comparing LNG-IUS with systemic medication or blank control among women diagnosed with primary dysmenorrhea or secondary dysmenorrhea (adenomyosis or endometriosis) from inception to 2020.04. Der Simonian-Laird random-effect model was used to pool data.P = 0.63, I2 = 4%) and other adverse outcomes. Sensitivity analysis regarding randomization methods was robust. Nine RCTs enrolled endometriosis women. Pooling results showed no significant difference between LNG-IUS and systemic medication treatment in terms of VAS at 6 months . Moreover, LNG-IUS was associated with higher risk of irregular vaginal bleeding (26.8% vs. 0).Seventy-one RCTs (6551 patients) were included. Overall bias risk was medium. Sixty-two articles enrolled patients with adenomyosis; LNG-IUS significantly reduced the visual analogue scale (VAS) score compared with the systemic medication group among adenomyosis women at 3 months (standardized mean difference (SMD) = −0.81, 95% confidence interval (CI) −1.22 to −0.40); 6 months ; 9 months ; 12 months . No difference was found in the incidence of irregular vaginal bleeding (16 RCTs; RR = 0.91, 95%CI: 0.62−1.33, LNG-IUS was associated with a reduced severity of dysmenorrhea compared with systemic medication; it was also beneficial for better control of menstrual blood loss and fewer adverse outcomes. Owing to small sample sizes, further well-designed RCTs are warranted to confirm these findings and long-term effects of LNG-IUS in the treatment of dysmenorrhea.https://www.crd.york.ac.uk/prospero/, identifier: CRD42021228343. Dysmenorrhea is the most common gynecologic disorder, and it affects about 60%–90% women of reproductive age , 2. It iin vivo at a rate of approximately 17.5 mcg/day after 24 days. This rate decreases progressively to 9.8 mcg/day after 1 year and to 7.4 mcg/day after 5 years. Levonorgestrel belongs to the class of medications called progestins, which is a hormone produced by the ovaries. This is a soft, flexible T-shaped contraceptive (birth control) device that is placed inside the uterus (intrauterine device or IUD). The medication is continuously released over a period of 3 years to prevent pregnancy. LNG-IUS was first designed as a contraceptive in 1970; it reduces the experience of pain and avoids the need to regularly take long-term medication. In recent years, LNG-IUS has also been officially approved for the treatment of heavy menstrual bleeding (The levonorgestrel-releasing intrauterine system (LNG-IUS) is an IUD. And LNG-IUS has three models, respectively are Mirena, Kyleena, Jaydess. But only Mirena went into public in China. Mirena releases 20 μg of levonorgestrel per day into the uterine cavity for a period of 5 years. Kyleena contains 19.5 mg of levonorgestrel (LNG) released bleeding . But in bleeding , the resWe performed a systematic review with meta-analysis using the protocol registered with PROSPERO (CRD42020203343).We included RCTs comparing LNG-IUS with systemic medication or blank control among women diagnosed with primary dysmenorrhea or secondary dysmenorrhea (adenomyosis or endometriosis). We excluded studies that did not report on our interested outcomes and those that were not published in English or Chinese. The primary outcome was defined as the degree of dysmenorrhea  at 3, 6, 9, and 12 months, and blood loss  at 3, 6, 9, and 12 months. The secondary outcome was defined as quality of life  and adverse events .We searched PubMed, EMBASE, the China National Knowledge Infrastructure (CNKI) and Wanfang Data, from inception to April 2020, with language limitation to English and Chinese. We used MeSH/Emtree terms and free-text to generate the search strategy, which comprised the terms “levonorgestrel releasing intrauterine system”, “LNG-IUS”, “dysmenorrhea”, “painful menses”, “menstrual pain”, “chronic pelvic pain”, and “dyspareunia”. The details of our search strategy are listed in Two paired reviewers who were well-trained in research methodology, independently screened titles, abstracts, and full texts for eligibility; assessed risk of bias; and collected data from eligible studies, using standardized, pilot tested forms with detailed instructions. Reviewers resolved disagreement through discussion or through adjudication by a third reviewer (LL) if agreement could not be reached.We used the Cochrane Handbook 5.1.0 bias assessment tool , 12 to a•Study characteristics: first author, country, publication year, study design, inclusion and exclusion criteria, number of study sites, sample sizes, length of follow-up.•Patient characteristics: age, parity•Intervention: details of LNG-IUS and systemic medication of the control group•Outcomes: degree of dysmenorrhea (VAS score), blood loss volume (PBAC score), quality of life, side effects .The following information from all eligible studies were collected:I2 and χ2 test. If I2 > 50% or χ2 test indicated P < 0.1, we defined it as statistical significance and we used the DerSimonian-Laird random-effect model to pool data. Subgroup analysis was undertaken to explore the potential heterogeneity based on systemic medication types , status of surgery and patient type (primary dysmenorrhea patients vs. secondary dysmenorrhea patients). Sensitivity analysis was also conducted by excluding studies with improper randomization sequence generation or unspecified randomization methods and excluding studies with sample size less than 50. Data analyses were undertaken by Review Manager 5.3.5. Publication bias was examined by Egger’s test and funnel plots.For dichotomous outcomes, we calculated risk ratio (RR) and their 95% confidence intervals (CIs). For continuous outcomes, we calculated mean difference (MD) or standardized mean difference (SMD) and their 95% CIs. Statistical heterogeneity was examined by Among 2641 identified publications, reviewers selected 71 RCTs , 78–84, n = 64), Brazil (n = 3), Egypt (n = 1), Thailand (n = 1), Turkey (n = 1), Taiwan (n = 1). Seventy studies were single center clinical trials, and one study ; 6 months , P < 0.0001; I2 = 94%); 9 months , P < 0.0001, I = 0); and 12 months , P < 0.0001, I2 = 94%; Of the 71 RCTs included, 62 –81, 86 eP = 0.0004, I2 = 36%) and 6 months , P < 0.0001, I2 = 96%). However, no difference was identified between these groups at 9 months  and 12 months , 6 months , P < 0.0001, I2 = 96%), and 12 months , P < 0.0001, I2 = 96%; Nineteen RCTs , 69, 86 P > 0.05). Li (P < 0.001).As for quality-of-life (QoL) assessment, Feng  reported.05). Li  reportedP = 0.63, I2 = 4%), headache , acne , or amenorrhea  between two groups. The systemic medication group was associated with a higher risk of nausea  and menorrhagia , while no significant difference was found in subgroup analysis according to whether the patients received surgery or not  after 6 months of treatment.Subgroup analysis by type of systemic medications showed a significant difference in VAS score  and 12 months . Nelsilene (P = 0.04) and self-image (P = 0.04) when compared to an etonogestrel implant group. Prasong (P < 0.05) but there was no difference in psychological domains (P = 0.229).As for QoL assessment, Petta  used theelsilene  reported Prasong  used theP = 0.07, I2 = 0%), nausea , headache , weight gain , or amenorrhea  between the two groups  and menorrhagia  in adenomyosis. We could not draw any conclusions according to available data because few studies  to explore the source of heterogeneity, which also brought limitations to the reliability of our results. Forth, we did not include “endometriosis” or “adenomyosis” as search words because we think dysmenorrhea was probably mentioned as their symptoms, which may have greatly limited finding valid studies for the research focus.Above all, there is still a need for more rigorous, well-designed and high-quality RCTs to investigate the long-term effect of LNG-IUS to confirm the current findings and to examine other vital outcomes such as number of dropouts and withdrawal of treatment.In conclusion, our systematic review demonstrated that the use of LNG-IUS was associated with a reduced severity of dysmenorrhea compared with systemic medication. Moreover, LNG-IUS was beneficial for improved control of menstrual blood loss and fewer adverse outcomes. At the same time, further well-designed RCTs are warranted to confirm the current findings and long-term effect of LNG-IUS in the treatment of dysmenorrhea.