Healthy microbiota is a critical mediator in maintaining health and it is supposed that dysbiosis could have a role in the pathogenesis of a number of diseases. Evidence supports the hypothesis that maternal dysbiosis could act as a trigger for preterm birth; aberrant colonization of preterm infant gut might have a role in feeding intolerance and pathogenesis of necrotizing enterocolitis. Despite several clinical trials and meta-analyses, it is still not clear if modulation of maternal and neonatal microbiota with probiotic supplementation decreases the risk of preterm birth and its complications. Modern molecular methods in microbiological diagnostics have recently revealed that modifications in microbial communities in maternal, fetal, and neonatal microbiota play a pivotal role in human health and disease.In the last decades, dysbiosis has been linked to the origin of many diseases . EvidencIn this review, we summarize the role of maternal dysbiosis in the pathogenesis of preterm birth, and the role of neonatal dysbiosis in feeding intolerance, necrotizing enterocolitis, and late-onset sepsis. Furthermore, we highlight the new perspectives about microbiota manipulation with prenatal and postnatal probiotic supplementation.An exhaustive search for eligible studies was performed using PubMed, Embase, Medline, Cochrane Library, and Web of Science databases.The following subject MeSH headings were used: “Premature Birth” [MeSH], “Dysbiosis” [MeSH], “Fetal Membranes, Premature Rupture” [MeSH], “Infant, Premature” [MeSH], “Enterocolitis, Necrotizing” [MeSH], and “Probiotics” [MeSH].To be as comprehensive as possible, proper Boolean operators such as “AND” and “OR” were also included. Additional studies were sought using references in articles retrieved from searches.Search limits were set for randomized placebo-controlled clinical trials (RCT), written in English, involving only human subjects, and published between March 2009 and March 2019.Preterm birth, and its complications, are the main cause of death for children below the age of five worldwide.  Each yeaRupture of membranes prior to 37 weeks of gestation and before the onset of labor, defined as “preterm prelabor rupture of the membranes” (pPROM), occurs in 30% of spontaneous preterm births .The clinical management of pPROM is challenging and often controversial: on the one hand, prolongation of pregnancy enables fetal maturation; on the other hand, rupture of the membranes increases risk of infection and poor neonatal outcomes.The correlation between lower genital tract infection and an increased risk of both pPROM and preterm delivery has lately raised great interest in the pathogenesis of such infection-related mechanisms . HoweverLactobacillus species dominate the community structure of a healthy vagina [During pregnancy, only one or few y vagina . These ly vagina .Lactobacillus spp. abundance as an early risk factor for pPROM [A recent study that aimed to assess the vaginal microbial community structure using next generation sequencing found an increasing vaginal bacterial diversity with reduced or pPROM .Gardnerella vaginalis, Atopobium vaginae, and Veillonellaceae bacterium were associated with increased risk of early preterm birth. Similarly, an increased presence of Prevotella, Peptoniphilus, Streptococcus, and Dialister species was linked with pPROM [Using hierarchical clustering analysis of 16S rRNA gene sequencing, Ravel and colleagues classified the vaginal bacterial population into community state types (CSTs) . The CSTth pPROM .Lactobacillus species with altered levels of proinflammatory cytokines are hypothesized to play a pivotal role [Also, bacterial vaginosis might increase the risk of preterm birth and perinatal complications.  Despite tal role ,14.Despite the fact that human placenta has been traditionally considered as a sterile environment, recent studies revealed that it has a distinct microbiota that can be influenced by maternal conditions, irrespective of the type of delivery ,16. ConvUreaplasma, Mycoplasma, Aerococcaceae, Bifidobacteriaceae, and Fusobacteria are found predominantly or exclusively in membranes of preterm newborns, highlighting how bacterial ascension from the urogenital tract is the main cause of chorioamnionitis and amniotic inflammation. [Specifically, microbial species of mmation. Gut and vagina maternal microbiota, via hematogenous spread and ascending colonization, also influence the fetal microbiota composition throughout the pregnancy .Furthermore, increasing evidence suggests that the maternal microbiota, according to its tendency and capacity to gain electrons, influences the placental oxidative reduction potential and might have direct or indirect consequences on the oxidative balance. Hence, gut dysbiosis may both be as the cause as the consequence of increased levels of oxidative species, thus shedding light upon the interplay between the gut microbiota and the placenta in causing congenital malformations and prematurity ,21.Bifidobacteria, lactobacilli, and non-pathogenic fungi.It has been argued that probiotics administration during pregnancy and the postnatal period might decrease some maternal and neonatal adverse outcomes . ProbiotThe probiotics might confer protection against preeclampsia, gestational diabetes, vaginal infections, maternal and infant weight gain, and diseases later in life .Preeclampsia still accounts for a vast number of deaths and severe maternal conditions. It is likely to be the consequence of abnormal placentation and maternal inflammatory response . In thisThe Norwegian Mother and Child Cohort Study, a large prospective cohort study, revealed an association between consumption and timing of milk-based products with probiotic lactobacilli and decreased risk of preeclampsia ,26.Moreover, probiotics can reduce C-reactive protein, a marker of inflammation linked to maternal conditions such as preeclampsia and gestational diabetes .Lactobacillus rhamnosus GG and Bifidobacterium lactis Bb12) demonstrated that probiotic supplementation significantly reduced the frequency of gestational diabetes mellitus (13% to 36%) without adverse events in either mothers or newborns [Studies addressing the role of probiotics to improve glycemic status have shown contradictory results. One randomized placebo-controlled study of 256 healthy women treated with diet and probiotics . According to their analysis, the administration of probiotics during pregnancy did not affect the risk of preterm birth or other secondary outcomes, including gestational diabetes. The authors concluded that taking probiotics or prebiotics during pregnancy to prevent preterm birth or newborns and maternal adverse pregnancy outcomes is still not supported by evidence .Conversely, probiotics supplementation during pregnancy may help to prevent atopic dermatitis in children. A meta-analysis in 2015 reported data from 4755 children and showed a significantly decreased risk for eczema compared to controls, especially when probiotic supplementation occurred during pregnancy, lactation, and neonatal period. However no significant difference in terms of prevention of asthma, wheezing, or rhinoconjunctivitis was documented .Furthermore, another recent review and meta-analysis showed that probiotics during pregnancy may reduce risk of eczema in children .Supplementation of probiotics to the mothers during pregnancy and/or breastfeeding can also change the cytokine profile of mother’s milk, increase fecal IgA in the offspring , and couThe role of neonatal microbial colonization in healthy full-term “vaginally delivered” newborns is a well-known important factor in mucosal immunity, nutrient absorption and digestion, visceral sensing and motility, and finally energy regulation . In healStaphylococcus, Propionibacterium, Bifidobacterium, and Lactobacillus with the capacity to digest human milk oligosaccharides. With the transmission of maternal microbiota to the fetus and newborn being a pivotal mechanism in the development of the neonatal gut eubiosis, every factor perturbating this physiological process might be responsible for aberrant gut microbiota colonization, defined as dysbiosis.Healthy colonization, called eubiosis, is characterized by a dominance of Gram-positive Firmicutes such as Klebsiella pneumoniae and Clostridium difficile) with reduced microbial diversity and low levels of short chain fatty acids (SCFA) [Bifidobacterium, Bacteroides, and Atopobium [The microbiota of children born preterm usually show a prevalence of pathogenic bacteria  . Furthertopobium ,42.This aberrant colonization has been generally linked to absent or limited exposure to maternal microbiota due to shortened labor or cesarean birth , exposurA recent study showed an increased proportion of Gammaproteobacteria in very low birth weight (VLBW) preterm newborns, with vaginal birth and antenatal steroids identified as a predictor of its abundance .Evidence has shown how preterm infant colonization might be associated with adverse outcomes of prematurity, such as feeding intolerance, necrotizing enterocolitis , late-onEpithelial structure integrity and barrier function of intestinal enterocyte are influenced by host–microbiota interactions and in preterm infants, the aberrant colonization may be responsible for the structural disruption that increases intestinal permeability, with translocation of bacteria to the bloodstream as well as cell death and necrosis .Preterm immature gastrointestinal tract with a “leaky” barrier and aberrant motility, digestion, and absorption predispose the preterm infant to an increased risk of feeding intolerance ,53.Despite feeding intolerance usually represents a benign condition , its preKlebsiella in this population. Despite the fact that no specific microbial pathogen has been identified as being responsible for the development of NEC, authors speculate that Klebsiella might be considered as a diagnostic biomarker preceding the onset of feeding intolerance and NEC [In 2019, a case–control study investigated the gut microbial composition of preterm infants with feeding intolerance showing a significantly higher abundance of  and NEC ,63.Late-onset sepsis (LOS) is a major cause of morbidity and mortality in preterm infants due to the immaturity of the immune system in the hostile environment of NICUs ,65. ReceRecently, other available data speculate that neonatal colonization contributes to other short- and long-term sequelae of prematurity, such as respiratory distress and bronchopulmonary dysplasia (BPD), retinopathy of prematurity (ROP) , somaticUreaplasma species colonization was significantly associated with the development of BPD, a chronic lung disease of prematurity leading to life-long limitations in lung function [The role of neonatal microbial colonization of the respiratory tract, in utero and during birth, is a novel research field. Emerging data have highlighted that diverse microbial populations inhabit the respiratory tract, previously considered a sterile compartment . A recenfunction .Despite some authors highlighting the possible key role of dysbiosis in the pathogenesis of ROP and non-communicable chronic diseases, this evidence remains somewhat speculative, and other factors, such as respiratory support or diet, need to be further investigated ,75,76.Furthermore, evidence supports the hypothesis that intestinal microbiota in preterm infants and the consequent “leaky gut” may alter the systemic inflammatory response. In this view, circulating pro-inflammatory cytokines, oxidative stress, and endotoxemia, resulting in systemic low-grade inflammation and aberrant production of neuroendocrine–immune mediators may exert a direct effect on gut–brain axis with long-term neurodevelopmental impairment .The concept of managing the intestinal microbiota or correcting dysbiosis with probiotic supplementation is a promising research field in neonatology .Multiple randomized placebo-controlled clinical trials (RCT) and cohort studies of probiotics in premature infants have been performed.The main mechanisms for the benefits of probiotics in preterm infants include gut barrier enhancement, immune response modulation, and competitive inhibition of colonization with pathogens.The differences in microbial colonization between preterm infants and term infants and between preterm infants with NEC and without NEC suggest that interventions focused on modulation of the gut bacteria may play a role in preventing NEC.In the last decades, different meta-analyses have investigated the efficacy and safety of probiotic supplementation for preventing NEC ,87,88,89Bifidobacterium and Lactobacillus to prevent NEC and reduce mortality in preterm infants of <34 weeks’ gestation or with a birth weight <1500 g [Despite each meta-analysis having used different inclusion criteria and statistical approach, the evidence is strong regarding the use of combination probiotics of both  <1500 g ,91. More <1500 g .Other positive outcomes of probiotic supplementation seemed to be a shorter time to full enteral feeding , shorterTwo recent systematic reviews and a meta-analysis concluded that probiotic supplementation might reduce the risk of LOS in preterms <37 weeks gestation or with a birth weight <2500 g. The results were also significant in studies that included only infants with gestational age <32 weeks or birth weight <1500 g ,95. FurtCandida colonization in preterm neonates in NICU [Current evidence also indicates that probiotics can reduce the risk of  in NICU , despite in NICU .In 2017, a meta-analysis assessing the effect of probiotics on the pathogenesis of ROP demonstrated no significant effect of probiotic supplementation on the risk of developing ROP. Similarly, another meta-analysis assessing the possible role of probiotics in altering the risk of BPD showed no significant effect .Finally, probiotic administration was not beneficial for the prevention of intraventricular hemorrhage (IVH) .Probiotics supplementation is not without risk, particularly in vulnerable populations such as premature infants, where the chance of probiotic translocation and sepsis is higher due to compromised gut integrity. In the literature, cases of probiotic sepsis have been reported , and in The next frontier for managing dysbiosis in preterm newborns will be the use of post-biotics (metabolites and components from live or dead bacteria)  and paraA healthy microbial colonization (eubiosis) of the maternal vagina and the neonatal gastrointestinal tract seems to correlate with a better newborn state of health.On the other hand, aberrant or preterm gut microbiota colonization (dysbiosis) might be associated with adverse outcomes of prematurity such as feeding intolerance, necrotizing enterocolitis, and late-onset sepsis. Maternal dysbiosis could represent a risk factor for premature labor as much as preterm newborns dysbiosis could play a role in the state of health and in the pathogenesis of various diseases of prematurity.Managing dysbiosis and manipulating the microbial environment using probiotic supplementations is a promising research field for promoting health and preventing diseases, reducing the risk of preterm birth and its complications.In the future, the management of dysbiosis in prematurity could also be based on the use of post-biotics and para-probiotics which, together with probiotics, could provide an additional benefit for premature infants.Further studies are needed to support these hypotheses and to identify the specific probiotic strain to be used, useful dosage regimens, and the effect of specific probiotics in high-risk populations such as ELBW infants.


This study aimed to describe the variation in risks of adverse birth outcomes across ethnic groups and socioeconomic circumstances, and to explore the evidence of mediation by socioeconomic circumstances of the effect of ethnicity on birth outcomes.England and Wales.The data came from the 4.6 million singleton live births between 2006 and 2012.The main exposure was ethnic group. Socioeconomic circumstances, the hypothesised mediator, were measured using the Index of Multiple Deprivation (IMD), an area-level measure of deprivation, based on the mother’s place of residence.The primary outcomes were birth outcomes, namely: neonatal death, infant death and preterm birth. We estimated the slope and relative indices of inequality to describe differences in birth outcomes across IMD, and the proportion of the variance in birth outcomes across ethnic groups attributable to IMD. We investigated mediation by IMD on birth outcomes across ethnic groups using structural equation modelling.Neonatal mortality, infant mortality and preterm birth risks were 2.1 per 1000, 3.2 per 1000 and 5.6%, respectively. Babies in the most deprived areas had 47%–129% greater risk of adverse birth outcomes than those in the least deprived areas. Minority ethnic babies had 48%–138% greater risk of adverse birth outcomes compared with white British babies. Up to a third of the variance in birth outcomes across ethnic groups was attributable to differences in IMD, and there was strong statistical evidence of an indirect effect through IMD in the effect of ethnicity on birth outcomes.There is evidence that socioeconomic circumstances could be contributing to the differences in birth outcomes across ethnic groups. This study was based on a large number of observations, which provided sufficient statistical power to investigate associations across a variety of factors and subgroups.The study was population based and sufficiently large to enable individual ethnic groups to be studied.We used an area-level measure to quantify the socioeconomic circumstances of individuals, although we also conducted a sensitivity analysis based on an individual-level measure available in a subset of the population.Observational data such as these are prone to unmeasured confounding, thus despite the reporting of adjusted estimates, some potential for residual confounding by unmeasured factors remains.The rates of adverse birth outcomes, and indeed most other adverse health outcomes, have declined steadily over the last several decades all over the world.Despite these mostly positive trends, stark disparities in birth outcomes persist across some individual and group characteristics. Previous research has shown that worse socioeconomic circumstances,At the same time, past and recent patterns of migration in England and Wales and many other parts of Europe have resulted in increasing proportions of births to women from ethnic minority groups.Designing interventions and policies aimed at reducing health disparities requires an understanding of the mechanisms through which heterogeneity of outcomes arises.This was a retrospective cohort study of all singleton live births at gestational age of 22 weeks or more in England and Wales between 2006 and 2012.Data from the national birth and death registries in England and Wales have been linked since 1993. The Office for National Statistics (ONS) checks these data to minimise inconsistencies. However, these data do not include characteristics such as ethnic group and gestational age. To obtain these characteristics, the registries data have been further linked to the birth notifications made by midwives and birth attendants using the babies’ National Health Service (NHS) numbers. Until 2015, birth notifications were done using a system called the NHS Numbers for Babies (NN4B).Outcomes of interest were: neonatal death, defined as a death from any cause occurring within the first 28 days of life; infant death, defined as a death from any cause occurring before the first birthday; and preterm birth, defined as a birth occurring before 37 completed weeks of gestation.The main explanatory variable was ethnic group of the baby as recorded in the NN4B notification system, based on the list of ethnic categories used in the 2001 Census in England and Wales. The development of this list has been described elsewhere.Socioeconomic circumstances were considered a potential mediator of the association between ethnicity and birth outcomes. It was assessed using the 2015 English Index of Multiple Deprivation (IMD)An individual-level measure of socioeconomic circumstances, the National Statistics Socioeconomic Classification (NS-SEC), was available for a 10% sample of live births in which information on parental occupation recorded at birth registration was coded by the ONS. The NS-SEC is an occupationally based measure of employment relations and conditions of occupations, conceptually relevant in showing the structure of socioeconomic circumstances and explaining various social phenomena.The linked data set also included covariates such as mother’s age, mother’s country of birth which was grouped into UK and non-UK for this analysis, birth registration type , year of birth between 2006 and 2012 and sex of child. These variables were included in our analyses as potential confounders of the association between socioeconomic circumstances measured using IMD and birth outcomes. Mother’s age and year of birth were modelled as continuous variables and any departures from linear association with the outcomes were explored and fitted.The slope index of inequality (SII) and the relative index of inequality (RII) were estimated to describe the differences in birth outcomes across IMD.c, the total effect, not shown in a), followed by estimation of the effect of ethnic group on birth outcomes (path c′) when adjusted for IMD (path b). Since the main exposure variable was multicategorical, the relative indirect effect of IMD was estimated for each ethnic group, with white British ethnic group as the comparison group,Two approaches were used to investigate the role of relative deprivation in the association between ethnic group and birth outcomes. First, the amount of variation in birth outcomes across ethnic groups which was explained by socioeconomic circumstances—measured using IMD deciles and modelled by linear regression—was estimated as the relative change in the residual ethnic group-level variance in outcomes as a function of relative deprivation.ab/c or equivalently 1−(c′/c). A lack of evidence of effect in path a or b implied the absence of an indirect effect; a lack of evidence of effect in path c implied that there was no effect to be mediated. In either case, the relative indirect effect was meaningless, therefore not calculated. Linear, rather than logistic, regression was preferred for modelling all effects despite the birth outcomes being binary variables because explained variance measures in linear models are most intuitive,The relative indirect effect, that is, the proportion of total effect mediated, was calculated as All analyses were of complete cases. Data management, manipulation and analyses were performed using Stata V.13.A patient and public involvement (PPI) group was formed to advise on the analysis plan, interpretation of results and dissemination plans for this study. We formed the group by contacting organisations who would potentially be interested in this project and asking if they would like to join the PPI group. The organisations which sent a representative to our PPI meetings were: Bliss, the Race Equality Foundation, Sands and the Lullaby Trust. In particular, the group provided guidance on ethnic groupings that were used in this study.Data on 4 744 666 infants were available in the linked data set. We sequentially excluded 16 695 infants with gestational age of 43 weeks and above, 20 999 infants with missing birth weights and 72 040 infants with implausible birth weights. A total of 4 634 932 infants with complete data in most covariates were included in the main analysis and 463 533 in the sensitivity analysis . There w10.1136/bmjopen-2018-028227.supp1Supplementary dataThe overall proportion of preterm births was 5.6%. Infant and neonatal deaths were less common, occurring in 3.2 and 2.1 per 1000 live births, respectively. However, more adverse birth outcomes occurred in infants born in deprived areas, overall  and withThe slope and relative indices of inequality in IMD across birth outcomes are presented in There was strong evidence of an indirect effect of ethnic group on birth outcomes through socioeconomic circumstances (Sobel test p<0.001). Infants of all ethnic groups except white non-British infants had higher adjusted risks of neonatal and infant mortality than white British infants . The samThese data from a cohort of babies born in England and Wales between 2006 and 2012 showed poorer birth outcomes among infants of non-white minority ethnic groups. Specifically, ethnic minority infants born over this period had up to twice the risk of adverse birth outcomes of white British infants. These data also showed that among all ethnic groups there was evidence of a gradient in birth outcomes across levels of socioeconomic circumstances, with infants born in the most deprived areas experiencing between 1.47 and 2.29 times the risk of adverse birth outcomes compared with those born in the least deprived areas. Additionally, minority ethnic infants were more likely to be born in deprived areas. Our study aimed to explore the evidence for relative deprivation as the driving mechanism for differences in birth outcomes across ethnic groups. This was based on a causal model in which we hypothesised that relative deprivation was a consequence of minority ethnic status, which was then a cause of poorer birth outcomes.We found that between a quarter and a third of the total between-ethnic group variance in birth outcomes was explained by relative deprivation. This finding, along with the evidence of the association between ethnicity and relative deprivation, and between relative deprivation and adverse birth outcomes, is consistent with a causal link between these factors, although not sufficient proof of causality. We also found strong evidence of an indirect effect through relative deprivation in the association between ethnic group and birth outcomes, with between 13.4% and 78.2% of the risk of adverse birth outcomes in non-white minority ethnic group infants estimated to be due to the higher levels of deprivation in these groups when compared with white British infants. Compared with white non-British infants relative deprivation mediated 8.8%–20.1% of the excess risk of adverse birth outcomes in white British infants. Together, these findings support the hypothesis that relative deprivation is one mediating mechanism through which observed differences in birth outcomes across ethnic groups arise. There may be other mediating mechanisms involving social, demographic or cultural factors, all of which are beyond the scope of this study, and this is further suggested by the finding of incomplete mediation across all groups and outcomes. The public health implication of these findings is that if we were to address the socioeconomic disadvantages borne by ethnic minorities in this population, we could reduce the between-ethnic group disparities in birth outcomes by between a quarter and a third in relative terms.These conclusions are further supported by previous studies which highlight the associations between ethnicity, relative deprivation and birth outcomes. For example, other studies have shown that birth outcomes and health of infants and children tend to worsen with greater area-level poverty,We hypothesise that the relationship between ethnicity and socioeconomic circumstances, and between socioeconomic circumstances and adverse health outcomes is causal. A number of established findings support this hypothesis of causality and highlight possible mechanisms of effect. For example, language and cultural differences between minority and majority groups can create barriers to accessing or benefiting from information.One limitation of this study is in the use of observational data to explore the causal relationships implied by the mediation model. Observational data are more prone to the occurrence of unmeasured confounding or reverse causality between the main exposure and the outcome, the mediator and the outcome, and between the exposure and the mediator. However, given that the exposure in this study was ethnic group, there were unlikely to be any common causes or reverse causality between it and relative deprivation—the mediator in this analysis—and with birth outcomes. And although there was unlikely to be any reverse causality between relative deprivation and birth outcomes, this association was likely to be confounded by unmeasured factors since both factors may have had external common predictors. In our analysis, we adjusted this pathway for mother’s age, mother’s country of birth, birth registration type , year of birth and sex of the child. Although this adjustment did not include factors such as education which may be common predictors of both relative deprivation and birth outcomes,These findings suggest that further reductions in adverse birth outcomes in England and Wales could be facilitated by interventions that reduce overall social inequality as this appears to be an important driver of variation in outcomes across ethnic groups. Such interventions must broaden their focus beyond health outcomes alone, and must aim to improve the totality of life circumstances among groups who experience the poorest outcomes.


Vm) for a conductive body that represents an unmyelinated axon (or the bare axon between the myelin sheath in a myelinated axon). We also computed the transmembrane potential of the sheath-covered axonal segment in a myelinated axon. We then systematically analyzed the biophysical factors that affect axonal polarization under transverse electric stimulation for both the bare and sheath-covered axons. Geometrical patterns of polarization of both axon types were dependent on field properties (magnitude and field orientation to the axon). Polarization of both axons was also dependent on their axolemma radii and electrical conductivities. The myelin provided a significant “shielding effect” against the transverse electric fields, preventing excessive axolemma depolarization. Demyelination could allow for prominent axolemma depolarization in the transverse electric field, via a significant increase in myelin conductivity. This shifts the voltage drop of the myelin sheath to the axolemma. Pathological changes at a cellular level should be considered when electric fields are used for the treatment of demyelination diseases. The calculated term for membrane polarization (Vm) could be used to modify the current cable equation that describes axon excitation by an external electric field to account for the activating effects of both parallel and transverse fields surrounding the target axon.Axonal stimulation with electric currents is an effective method for controlling neural activity. An electric field parallel to the axon is widely accepted as the predominant component in the activation of an axon. However, recent studies indicate that the transverse component to the axolemma is also effective in depolarizing the axon. To quantitatively investigate the amount of axolemma polarization induced by a transverse electric field, we computed the transmembrane potential ( Electrical stimulation of nerve cells was first reported by Luigi Galvani in 1780 , who accE//) and the other perpendicular  to the axon. The E// is widely regarded as the predominant factor that activates the axons . Here, m is the transmembrane potential and eV is the extracellular electric field applied to the fiber. It is nonzero only if E// is nonzero. The surface resistance and capacitance of the membrane are mR and mC, respectively. The intracellular resistivity is iR and the fiber radius is c. This simplification facilities the rapid calculation of neural activation. However, it ignores the presence of the cell, which perturbs the local extracellular electric field. It also ignores the mutual interactions between the neurons and the applied electric field  for an unmyelinated axon (or the node of Ranvier in myelinated axon), under transverse electric field stimulation. We also provide Vm for a myelin-covered axon, which was modeled as a co-centric, two shell structure. We investigate biophysical factors that may affect the Vm, including field intensity and orientation, axonal biophysics and the impact of pathological demyelination. We discuss the possibility of placing Vm into a modified cable equation, so that the contributions of both the longitudinal and transverse components of an electrical field could be simultaneously evaluated during electric stimulation of axons.In the present paper, we model a bare axon as a conductive cylindrical shell, and provide an analytical expression of membrane polarization . The axon was exposed to a transverse, direct current (DC) electric field (E0). It included a total of five isotropic and homogenous regions: the medium (#0), the myelin sheath (#1), the periaxonal space (#2), the axolemma (#3), and the axonal cytoplasm (#4). The dielectric permittivities and conductivities in the five regions were ε0, ε1, ε2, ε3, ε4 and σ0, σ1, σ2, σ3, σ4, respectively. The myelin had an outer radius (a), inner radius (b), and the thickness (b–a). The axon had an outer radius (c) and inner radius (d). Thickness of the axolemma was therefore c–d. We modeled a bare axon and its myelin sheath using homogeneous cylindrical volume conductors . Figure r, θ, z), the electric field distribution was calculated byUsing the cylindrical coordinates , it is written asThe potential, nt media . In a cyr = a),r = b),r = c),r = d),S0 = σ0 + jωε0, S1 = σ1 + jωε1, S2 = σ2 + jωε2, S3 = σ3 + jωε3, S4 = σ4 + jωε4. (C) Electric fields an infinite distance away should not be perturbed by presence of the axon. (D) The electric potential inside the cytoplasm (r = 0) was finite.Several boundary conditions were evaluated in solving the equation (Appendix): (A) The electric potential was continuous across the boundary of the two different media. (B) The normal current density was continuous across the two different media. “Complex conductivity,” defined as material . Here, σn) of lamella . Numerical simulations were performed with Matlab 8.4.0 .nA, nC were unknown coefficients . These coefficients were solved in the Appendix , we obtained the expression of voltage inside the axolemmaVm) of the field was obtained by subtracting the membrane potential at the inner surface from that of the outer surface of the axon (The solution for Laplace’s equation  was writ regions (9)\docuAppendix , by consthe axon , 2011, \red axon ,(11)\doVm and Φ were functions of both field properties and tissue properties. The field properties included the orientation of the field and its intensity. The tissue properties include the electric parameters (conductivity and di-electricity) and the geometrical parameters . The above Vm expression for the myelin-covered axon (S1 = S0 and S2 = S0 (Voltage drop (Φ) across the myelin sheath was obtained by subtracting the myelin potential at the inner surface from the outer surface of the myelin (12)\docred axon  was furt S2 = S0 ,(13)\doVm is determined by the axon’s orientation to the electric field. The axolemma should be hyperpolarized wherever an electric current enters the membrane and be depolarized wherever the current extrudes from the membrane  of the axon, and its conductivity. For the parametric analysis, we plotted the maximum polarization , in agreement with the literature that studied spherical cell polarization in an electric field (−3 S/m) due to membrane disruption and leakage, such as during electroporation .We first studied how the loss of myelin layers could affect rization . The traVm in a transverse electric field. When myelin conductivity was as low as 5 × 10−5 S/m, reduction in myelin thickness did not lead to dramatic changes in Vm , the axon could be significantly depolarized at any myelin thickness  under transverse field stimulation. It analyzes the biophysical factors that affect axonal polarization under physiological conditions, and under pathological conditions such as demyelination. Finally, it provides the needed term to modify the current cable equation, so that the equation can account for the effects of more realistic field, which include both the transverse and parallel directions.The model shows that the longitudinal axon was polarized with a distinct geometrical pattern by the transverse electric field, which was dependent on the orientation of the axon in the field. Previously, regional polarization has been observed in a variety of modeling and experimental studies for cells under electric  and magnVm was dependent on the intrinsic tissue properties of an axon. We observed that Vm was greater in larger diameter axons than in smaller ones (Vm) to the transverse field   . It is tductive)  could beDynamic changes of myelin also occur during development , neural Vm term could potentially be used to modify the current cable equation to account for both E// and E⊥. modified activating function, where E⊥ is more important.The analytical expression of the 2c=Rm4Ri . ComparaVm term for the unmyelinated axon  for a myelin-covered axon and a bare, unmyelinated axon, under a transverse DC electric field. Results show that the myelin sheath shields the axon from extensive depolarization. Demyelination could alter axon’s sensitivity to a transverse electric field if the process of demyelination involves significant increases in the electric conductance of the myelin. The analytical solution of Vm for the unmyelinated axon can be used to improve the activation function of the current cable equation that describes electric stimulation.This work provides novel analytical expressions of the electrically-induced transmembrane potential (V was bounded at r = 0 (boundary condition D), C4 = 0.At an infinite distance, according to boundary condition (C), Expressions for the potential distribution in the five modeled regions were:V (from The  ∇V from  were conr = a)On the #0#1 interface (r = b)On the #1#2 interface (r = c)On the #2#3 interface (r = d)On the #3#4 interface (We solved  to A-13A-13 to o10.7717/peerj.6020/supp-1Supplemental Information 1Click here for additional data file.10.7717/peerj.6020/supp-2Supplemental Information 2Click here for additional data file.10.7717/peerj.6020/supp-3Supplemental Information 3Click here for additional data file.10.7717/peerj.6020/supp-4Supplemental Information 4Click here for additional data file.10.7717/peerj.6020/supp-5Supplemental Information 5Click here for additional data file.10.7717/peerj.6020/supp-6Supplemental Information 6Click here for additional data file.


The neonatal mortality rate (NMR) in Malawi has remained stagnant at around 27 per 1000 live births over the last 15 years, despite an increase in the uptake of targeted health care interventions. We used the nationally representative 2015/16 Demographic Health Survey data set to evaluate the effect of two types of maternal exposures, namely, lack of access to maternal or intra-partum care services and birth history factors, on the risk of neonatal mortality.A causal inference approach was used to estimate a population attributable risk parameter for each exposure, adjusting for co-exposures and household, maternal and child-specific covariates. The maternal exposures evaluated were unmet family planning needs, less than 4+ antenatal care visits, lack of institutional delivery or skilled birth attendance, having prior neonatal mortality, short (8-24 months) birth interval preceding the index birth, first pregnancy, and two or more pregnancy outcomes within the preceding five years of the survey interview.P < 0.001), while non-institutional deliveries and the shortest preceding birth interval (8-24 months) each had an attributable risk of 1.3/1000 (Ps = 0.01). Having 2 or more pregnancy outcomes within the last 5 years had an attributable risk of 3/1000 (P = 0.006). Attending less than 4 ANC visits had, a relatively large attributable risk , and it was not statistically significant at alpha level 0.05.We included 9553 women and their most recent live birth within 3 years of the survey. The sample’s overall neonatal mortality rate was 18.5 per 1000 live births. The adjusted population attributable risk for first pregnancies was 3.9/1000 (Our analysis addresses the gap in the literature on evaluating the effect of these exposures on neonatal mortality in Malawi. It also helps inform programs and current efforts such as the Every Newborn Action 2020 Plan. Increasing access to maternal care interventions has an important role to play in changing the trajectory of neonatal mortality, and women who are at an increased risk may not be receiving adequate care. Recent studies indicate an urgent need to assess gaps in service readiness and quality of care at the antenatal and obstetric care facilities. The neonatal mortality rate (NMR) in Malawi has remained stagnant at around 27 per 1000 live births over the last 15 years . This isStudies using summary data at national levels have shown that improved program efforts can lead to a reduction in preventable neonatal mortality rates in low-income countries in order to make them comparable to those in more industrialized countries. Bhutta et al 2014  used theIn addition to health care interventions or service related factors, there are maternal and child biological and socio-demographic factors, as well as household risk factors that have been well documented. For example, male sex is associated with an increased risk, and neonatal mortality tends to cluster within a small subset of mothers and those who have high fertility -24. ExamMost evaluations of the effects of health care interventions for reducing NMR in Malawi have been done on a small scale within a few districts, or focusing on only one or two interventions ,10,11,31The aim of this study was to add to the literature using more recent data from the Malawi 2015/16 DHS. We applied a causal inference approach to individual mother-level data to evaluate the relative effects of two types of risk exposures, namely, lack of access to maternal care and intra-partum services and prior birth history events, on the risk of neonatal mortality. The causal inference approach used here enables us to adjust the attributable risk estimate of each exposure for the effect of its co-exposures, as well as household, maternal and child-specific covariates. The results of this analysis can help guide current efforts to reduce neonatal mortality such as the Every Newborn Action Plan  and to hData about each participating woman’s most recent pregnancy and live birth (within 3 years leading to the survey) were extracted from the Malawi DHS 2015/16. Thus, all information was based on the woman’s ability to recall the health care services accessed and their birth history. Various combinations of the following risk exposures were evaluated  and region . The DHS followed a two-stage stratified sampling design where each of the 28 districts in Malawi was stratified into urban or rural, and within those strata, the standard enumeration areas were sampled proportional to size.multiPIM by Ritter et al (2014) , averaged over the covariates:Under the causal inference assumptions , the PIMjth a exposure φj*  is the amount of the outcome that would have been averted if everyone was unexposed to exposure Aj*. In that sense, the PIM parameter is the reverse of the population attributable risk which is traditionally used in in epidemiology studies.In other words, for a P associated with the test of the null hypothesis that its true value is 0. One of the main advantages of following this approach that is worth noting here is the flexibility to specify different candidate parametric and non-parametric models for estimating g(0|W) and Q. The best among these is selected via v-fold cross-validation. This is referred to as the ‘super learner’ approach. We estimated our TMLE parameters with logistic regression for the exposure (g) models and a nonparametric recursive partitioning for the outcome (Q) models.In the derivation of the potential effects of each exposure, we adjusted for other co-exposures in addition to household, mother and child covariates as well as the DHS sampling weights in the intervention model g(0|W). We present the φ parameter alongside its estimated standard error and the Analysis was carried out with 3 combinations of the risk exposures due to a strong correlation (hence complete confounding) between birth intervals and the number of pregnancy outcomes in the past 5 years, and between institutional delivery and skilled birth attendance; and also because of small sample sizes in the high risk categories of previous neonatal mortality and shortest birth intervals. With small sample sizes, different stratifications led to some categories having probabilities that were completely determined .The exposures that had the least proportions of exposed mothers during their most recent pregnancy were ANC4+ visits (50%), two or more births under 5 in the last 5 years (36%), first pregnancy(26%). Institutional delivery / SBA and the shortest birth interval of 18-24 months had the least number of exposed women , first pregnancies were associated with the highest-risk of 3.9 per 1000 live births (P < 0.001). Having less than 4 ANC visits had the second largest attributable risk of 2.1 per 1000 live births because of the high prevalence of this exposure but this was not statistically significant after adjusting for covariates and co-exposures. Non-institutional delivery and the shortest birth interval (8-24 months) each had a population attributable risk of 1.3 per 1000 live births (Ps = 0.01). In a sensitivity analysis where SBA was used instead of institutional delivery, the population attributable risk for non-SBA was 1.6 per 1000 live births (P = 0.005). In a sensitivity analysis where first pregnancy was added without the shortest birth interval, the population attributable risk for first pregnancy was 2.9 per 1000 live births (P = 0.02).P = 0.004). In Exposure Set C, having previous neonatal mortality had a negligible population attributable risk 0.2 per 1000 live births (P = 0.34). In all three models, having unmet family planning needs for spacing or limiting was found to have an attributable risk of 1.2 /1000 (Ps = 0.08). In sensitivity analyses where birth history was excluded, the population attributable risk of having family planning needs unmet remained the same.In Exposure Set B, first pregnancy and birth interval were removed and replaced by having 2 or more pregnancy outcomes within the last 5 years of the survey and this had a population attributable risk of 3 per 1000 live births  models and the outcome (Q) models.One of the limitations of our study is that our data on access to different interventions and birth history services is dependent on the participating mother’s ability to recall the events. We limited this by only evaluating data from the most recent births within 3 years of the survey. Another potential problem is that there may be misclassification of neonatal deaths and stillbirths . The mulOur analysis addresses the gap in the literature on evaluating the effect of these exposures in Malawi, and it helps inform programs and current efforts such as the Every Newborn Action 2020 Plan. It shows that increasing access to maternal care interventions has an important role to play in changing the trajectory of neonatal mortality. Birth history factors play an important role on the risk of NM and we hypothesize that women who are at an increased risk may not be receiving adequate care. Recent studies indicate that gaps still exist in service readiness and quality of implementations of the existing WHO-recommendations for family planning and antenatal care services, as indicated by recent studies.


The preterm birth rate is rising in high-income countries and is associated with increased mortality and morbidity. Although the risks increase with greater prematurity and risk factors have been found to vary with gestational age and labour onset, few studies have focused on the myriad pathways to extreme preterm birth (20–27 weeks’ gestation). The current study investigated trends in extreme preterm birth by labour onset type and examined the antecedent risks to further our understanding around the identification of high-risk pregnancies.Retrospective cohort study including all singleton extreme preterm births in Western Australia between 1986 and 2010. De-identified data from six core population health datasets were linked and used to ascertain extreme preterm births  after spontaneous onset of labour, preterm pre-labour rupture of membranes, and medically indicated labour onset. Trends over time in extreme preterm birth were analysed using linear regression. Multivariable regression techniques were used to assess the relative risks associated with each salient, independent risk factor and to calculate Population Attributable Risks (PARs).The extreme preterm birth rate including medical terminations and birth defects significantly increased over time whereas the extreme preterm birth rate excluding medical terminations and birth defects did not change. After medical terminations and birth defects were excluded, the rate of medically indicated extreme preterm births significantly increased over time whereas the rate of preterm pre-labour rupture of membranes extreme preterm births significantly reduced, and the rate of spontaneous extreme preterm births did not significantly change. In the multivariate analyses, factors associated with placental dysfunction accounted for >10% of the population attributable risk within each labour onset type.First study to show that the increase in extreme preterm birth in high-income jurisdiction is no longer evident after medical terminations and birth defects are excluded. Interventions that identify and target women at risk of placental dysfunction presents the greatest opportunity to reduce extreme preterm births. Research has demonstrated that the preterm birth (< 37 weeks gestation) rate is rising in high-income countries –4. PreteA complex and inter-related range of risk factors for preterm birth have been identified in the extant literature. These include distal characteristics associated with sociodemographic circumstances, genetic traits, reproductive history and some maternal medical conditions, and factors that are more proximal to the birth—such as complications of pregnancy and delivery –9. An emAntecedent risk profiles have been shown to differ by labour onset types, reflecting the myriad different pathways to preterm birth. The evidence-base, to-date, is derived from a handful of studies that have investigated the trends and predictors of all preterm births after spontaneous onset of labour, pre-labour rupture of membranes, and medically indicated labour onset , 13–16. This retrospective cohort study included all singleton term and extreme preterm births in Western Australia between 1986 and 2010, inclusive. Study data were sourced from core population health datasets held by the Data Linkage Branch of the Western Australian Government Department of Health (DLB). These datasets include the Midwives’ Notification System (MNS), Western Australian Register of Developmental Anomalies (WARDA), the Birth Registration and Death Registration datasets , the Hospital Morbidity Data Collection (HMDC), and the Mental Health Information System (MHIS).The MNS records the circumstances of all births of 20 weeks or more gestation, with information received from attending midwives. The WARDA includes all birth defects diagnosed at birth for stillbirths (including terminations of pregnancy) and livebirths as well as diagnoses for livebirths up to six years of age from a number of sources with a high level of ascertainment . The HMDThese data were linked together by the DLB by probabilistic linkage using common identifiers including name, address and birthdate . MultiplThis research was granted ethics approval by the Western Australian Department of Health Human Research Ethics Committee (#2011/64) and the Western Australian Aboriginal Health Ethics Committee (#613). These ethical approvals support a waiver of consent on the basis that the study: (1) utilises routinely collected information from existing administrative datasets ; and (2) only has access to de-identified data, which are stored, analysed and disseminated according to strict protocols.Consistent with World Health Organization guidelines , extremeWe included data on a range of known and available risk factors that typically occur prior to 20 weeks gestation, including sociodemographic, reproductive history, maternal conditions, and pregnancy complications. Sociodemographic data included socio-economic status, geographic isolation, maternal age, marital status, maternal ethnicity and sex of the child. Socio-economic status at the time of the birth was assessed using the Socio-Economic Index for Areas (SEIFA) index of relative socio-economic disadvantage. The SEIFA index is calculated by the Australian Bureau of Statistics based on population census data relating to householder’s education, occupation, employment, income, housing and household composition . The levMaternal ethnicity was sourced from the MNS. Maternal self-report of ethnic origin has been categorised by the attending midwife into the categories of Caucasian, Aboriginal and/or Torres Strait Islander , Asian , Indian (Indian subcontinent), African, Polynesian and Maori. The number of births to Polynesian women was very small and, accordingly, the maternal ethnicity of these cases was recoded and included in the group ‘Other’ which includes ethnicities not covered by the above . MaternaMaternal reproductive history data sourced from the MNS included the number of previous pregnancies, stillbirth(s), preterm birth(s), and caesarean section(s). Maternal conditions included essential hypertension, maternal asthma, genital herpes, smoking, pre-existing diabetes mellitus, overweight/obesity, substance abuse, and mental health diagnosis. Data relating to essential hypertension, maternal asthma, genital herpes, and smoking cigarettes during pregnancy were sourced from the MNS. Maternal overweight/obesity diagnosis  prior to the birth of the child was derived from the HMDC records when any of the following ICD diagnostic codes were recorded: ICD-10 E56 and E68.9, ICD-9 278, and ICD-8 277 and 278. Maternal substance abuse diagnosis  prior to the birth of the child was derived from the HMDC and MHIS records when any of the following ICD diagnostic codes were recorded: ICD-10 F10 to F19 and F55, ICD-9 291, 292, and 303 to 305, or ICD-8 291, 294.3, 303 and 304. Maternal mental health diagnosis  prior to the birth of the child was derived from the HMDC and MHIS records when any of the following ICD diagnostic codes were recorded: ICD-10 F00 to F09, F20 to F54, F56 to F99, and X60 to X84, ICD-9 290, 293 to 302, 306 to 319, and E950 to E959, or ICD-8 290, 292 to 294.29, 294.4 to 302, 305 to 319, and E950 to E958. ICD diagnostic codes were selected to be as inclusive as possible but it should be noted that, as they relate to hospital admissions, they likely reflect only the most serious cases of maternal overweight/obesity, maternal substance abuse, and maternal mental health problems.Pregnancy complications data sourced from the MNS included threatened abortions, urinary tract infection, pre-eclampsia, antepartum haemorrhage , threatened preterm labour, fertility treatment, chorionic villus sampling/placental biopsy, amniocentesis, and year of birth. MNS data regarding gestational age and birthweight were used to calculate a measure of growth restriction: growth/weight for gestational age using Australian weight for gestational age norms .Information on pregnancy terminations at 20 weeks or more gestation was obtained from the WARDA and the cause of death text field on the Death Registration record. In Western Australia, all termination procedures are notifiable under the Health Act 1911. While no upper gestational age limit is specified in the legislation, the vast majority of terminations are conducted prior to 20 weeks . Late teThe WARDA was used to identify children with birth defects, and codes cases according to the British Paediatric Association extension of the International Classification of Diseases Version 9. All cases with a birth defect diagnosis code(s) were classified as having a birth defect for the purposes of this study. A full list of birth defects diagnosis codes is available on the WARDA website .βs) are reported along with absolute changes in rates (based on unstandardised regression coefficients) where significant changes are observed). The assumptions for linear regression were assessed by visual inspection of scatterplots for linearity and by using normal probability plots of the standardised residuals along with scatterplots of the standardised predicted value versus the standardised residuals.Trends over time in the extreme preterm birth rate were analysed using linear regression, with year of birth specified as a (continuous) classification variable and the t-statistic used to assess statistical significance  after excluding medical terminations and cases with birth defects (spontaneous term births were used as the comparison group because they represent the preferred outcome). Risk factors that had a significant risk ratio at the univariable level were included in the relevant multivariable model, with the exception of birth year group, which was included in all models to control for changes in the effects of risk factors over time. Adjusted estimates of relative risk were derived using a modified Poisson regression model  with a robust error variance . For eacβ = 0.84, p < .001; an increase of 0.10 per 1,000 births per year), the extreme preterm birth rate excluding medical terminations demonstrated a non-significantly increase over time , and the extreme preterm birth rate excluding medical terminations and birth defects did not change over time .For the 1986–2010 period there were 638,463 singleton births  including 4,202 extreme preterm (20–27 weeks gestation) births . After excluding medical terminations, there were 637,696 singleton births  including 3,493 extreme preterm births . After excluding medical terminations and cases with birth defects (minor and major), for the 1986–2010 period there were 603,351 singleton births  including 2,997 extreme preterm births . As can be seen in β = 0.94, p < .001; an increase of 0.10 per 1,000 births per year) whereas the rate of spontaneous extreme preterm births  and preterm pre-labour rupture of membranes extreme preterm births  did not significantly change over time.The incidence of spontaneous, preterm pre-labour rupture of membranes and medically indicated extreme preterm births including medical terminations and birth defects are displayed in β = 0.63, p = .001; an increase of 0.02 per 1,000 births per year). Thus, the majority of the increase (around 80%) in the rate of medically indicated extremely preterm births is explained by medical terminations and birth defects .N = 150,584). Descriptive statistics (% of births) and the results of univariate logistic regression analyses are shown in Data on the full range of risk factors was available for 1998–2010 (3.7% of cases had missing data on one or more variables). For this period, excluding medical terminations and cases with birth defects, complete sets of data were available for 629 cases of spontaneous extreme preterm birth, 496 cases of extreme preterm birth with preterm pre-labour rupture of membranes, and 430 cases of medically indicated extreme preterm birth. Risks are compared to those for spontaneous term births –placenta praevia, APH–placental abruption, APH–other, threatened preterm labour, fertility treatment, and amniocentesis. Whereas having fewer than 4 previous pregnancies was a significant protective factor, at the univariate level, against spontaneous, preterm pre-labour rupture of membranes and medically indicated extreme preterm birth.Most other risk factors were only associated with one or two of the extreme preterm birth labour onset groups at the univariate level. Essential hypertension was a significant risk factor, at the univariate level, for preterm pre-labour rupture of membranes and medically indicated extreme preterm birth. Being in the most disadvantage SES quintile was a significant risk factor for spontaneous and preterm pre-labour rupture of membranes extreme preterm birth as was having a urinary tract infection. African maternal ethnicity was a significant risk factor for preterm pre-labour rupture of membranes and medically indicated extreme preterm birth. Being small or large for gestational age had significant impact, at the univariate level, on the risk of spontaneous and medically indicated extreme preterm birth see . Other sEach multivariable model included risk factors that achieved statistical significance in univariable analysis see . At the The other significant risk factor for preterm pre-labour rupture of membranes extreme preterm birth was maternal age of 40 years or more. Other significant risk factors for medically indicated extreme preterm birth were essential hypertension, pre-eclampsia, being small or very small for gestational age, and being born later in the study period. In contrast, having 1 previous pregnancy was significantly protective against medically indicated extreme preterm birth as was being large for gestational age.There were differences in the factors with the highest PAR values for each labour onset type. The categories of fetal growth restriction collectively accounted for the highest proportion of the PAR for medically indicated extreme preterm births: 22.6% (95% CI: 18.2–26.9%) for growth restriction categorised as very small for gestation, and 12.2% (95% CI: 8.4–16.0%) for small for gestation. A notable PAR was also recorded for pre-eclampsia  with respect to medically indicated extreme preterm births. APH–other  was the only factor with a PAR of over 10% for preterm pre-labour rupture of membranes extreme preterm births. Threatened preterm labour , APH–other , primigravida , and the child being male  accounted for the largest proportion of the PAR for spontaneous extreme preterm birth.The current study sought to use linked Western Australian administrative data to examine the antecedent risks to extreme preterm births, with consideration of labour onset categories and the use of PARs. Overall, factors associated with placental dysfunction  represent the most prominent risks to extreme preterm birth, although individual risks associated with reproductive history, maternal conditions, pregnancy complications and socio-demographic circumstances also form an important part of the risk profile. This is consistent with the findings of previous research into the risk factors for preterm birth more broadly . TogetheConsistent with the findings of previous preterm birth research –15, we aResults revealed a significant increase in the extreme preterm birth rate over time when medical terminations and birth defects were included but no change when medical terminations and birth defects were excluded. This is consistent with our earlier work which found that changes in the medical termination rate, mainly related to improvements in the prenatal diagnosis of structural fetal anomalies, had a significant impact on the extreme preterm stillbirth rate across this period .After medical terminations and birth defects were excluded, the rate of medically indicated extreme preterm births was found to significantly increase over time , whereas the rate of preterm pre-labour rupture of membranes extreme preterm births demonstrated a significant reduction. The rate of spontaneous extreme preterm births did not significantly change over time. Future research should investigate the causes of the increase in medically indicated extreme preterm births and the decrease in preterm pre-labour rupture of membranes extreme preterm births with increased antenatal surveillance and changing maternal risk profiles likely to play a role see .A major strength of the current study is the use of linked administrative data enabling population-representative analysis that: (1) can account for medical terminations and birth defects in the examination of trends in birth rates, and (2) supports the simultaneous inclusion of a broad range of factors in examining antecedent risk profiles. However, the MNS excludes information on births that occurred before 20 weeks gestation. Because this is the period when the vast majority of medical terminations are performed , 36 one There are, typically, some data quality issues associated with the use of linked administrative data–these extend to the quality of linkage processes, changes in the methods of collection over time and between different sites, among others. Quality assessments consistently show that information from our source datasets are of high quality, low levels of missing information, and supported by best practice protocols that reduce missing links to negligible levels –21. AlsoThis study is the first to find that the increase in the extreme preterm birth rate observed in a high-income jurisdiction is no longer evident after medical terminations and birth defects are excluded. The findings underscore the role of pregnancy conditions related to placental dysfunction in the aetiology of extreme preterm birth. Identifying women at risk of placental dysfunction because of personal or family history and implementing effective interventions presents the greatest opportunity to decrease the population-level impact of extreme preterm birth.S1 Table(DOCX)Click here for additional data file.


To assess whether quality of maternal diet affects birth weight and the risk of small for gestational age (SGA) and/or large for gestational age (LGA) babies.z-scores, SGA, and LGA according to an ultrasound-based, population-based, and a customized growth standards. Associations were studied by linear and multiple logistic regression.This study is based on the Norwegian Mother and Child Cohort Study (MoBa) and includes 65,904 pregnant women who answered a validated food frequency questionnaire at mid-pregnancy. Three maternal dietary patterns were extracted based on characteristics of food items in each pattern. From these we created four non-overlapping groups: “high prudent,” “high Western,” “high traditional,” and “mixed”. We obtained information about birth weight from the Norwegian Medical Birth Registry and calculated birth weight βultrasoundz-scores −0.041 : −0.068, −0.013)) and the high traditional group with higher birth weight ) for all three growth standards. The high prudent pattern was associated with increased SGA risk (SGAultrasound odds ratio (OR) 1.25 ) and decreased LGA risk ), while the high traditional group on the contrary was associated with decreased SGA ) and increased LGA risk ).Compared to the high Western group, the high prudent group was associated with lower birth weight (Food quality was associated with birth weight in this well-nourished Norwegian population. Food quality may affect a woman’s risk of giving birth to a SGA or LGA baby. In about 20% of all births, the babies are born either small for gestational age (SGA) or large for gestational age (LGA) [Abnormal infant birth weight is correlated with increased infant mortality and morbidity –4. In ab4ge (LGA) . Intrautge (LGA) , hypoxiage (LGA) , neonatage (LGA) , and necge (LGA) . LGA is ge (LGA) . Moreovege (LGA) , accordige (LGA) –13.overall diet quality and birth weight and the risks of SGA and LGA are not well understood. Several studies have focused on individual foods or nutrients in relation to birth weight [quality and, to some extent also, the quantity of food [–Besides genetic potential  and plach weight , 18. How of food –21, withof food –. Some stof food –, 24, whiof food –. DietaryInformation about how maternal food quality is correlated to unfavorable birth weight could provide a basis for nutritional interventions to improve both maternal and infant health. The aim of this study was to examine associations between data-driven maternal dietary patterns and birth weight, SGA and LGA, based respectively on an ultrasound-based, population-based, and a customized growth standard, in a population-based pregnancy cohort.The Norwegian Mother and Child Cohort Study (MoBa) is a prospective, population-based pregnancy cohort study conducted by the Norwegian Institute of Public Health . ParticiThis study is based on two questionnaires, respectively, answered around gestational weeks 15 (Questionnaire 1) and 22 (Questionnaire 2). In Questionnaire 1, women reported, among other things, demographics, lifestyle, obstetric history , illnesses, health-related factors, physical activity, medications, and dietary supplements. In Questionnaire 2, they recorded average daily intake of food, beverages, and dietary supplements since the beginning of pregnancy. Questionnaire 2 is a semi-quantitative food frequency questionnaire (FFQ), developed and validated for use in MoBa . To be eInformation about birth weight and gestational age was obtained from the MBRN.SGA/LGA, ultrasound-based: >2 standard deviation (SD) above or below the expected birth weight for any given gestational age, according to Marsal’s ultrasound-derived growth curves .SGA/LGA, population-based: <10th or >90th percentile, according to gestational age-based growth curves derived from the Norwegian newborn population .z-scores were calculated according to the growth standards described above.SGA/LGA, customized: <10th or >90th percentile, as suggested by Gardosi et al. , is baseSince there is no consensus on growth standards, we assessed birth weight, SGA, and LGA according to the following three approaches:The MoBa FFQ was designed to cover maternal diet during the first 4–5 months of pregnancy. In a validation study with 119 MoBa participants, FFQ intakes were evaluated using four-day weighed food diaries and biological markers in blood and urine as reference measures . The FFQWe used a data-driven technique to empirically derive dietary patterns. Factor analysis with principal components is a technique used to reduce the dimension of the data by forming new linear combinations of the original observed variables based on correlation. Daily intakes (g per day) of foods and beverages in the 58 food groups were entered into the model and orthogonal (varimax) rotation was performed. Each participant is given scores on all factors (“patterns”), and the scores indicate the adherence of each individual to the respective dietary pattern. Based on visual examination of a scree plot, three dietary patterns were identified and respectively denoted prudent, Western, and traditional, based on the constituent foods. More details can be seen in Supplementary Table 2). Only women who reported weight between 35 and 180 kg and height above 1.40 m were included. Maternal height was divided into quartiles and used as a categorical variable . Data on parity were collected from the MBRN as well as from MoBa and were divided into two categories: nulliparous or parous. Information about smoking habits during pregnancy was collected from Questionnaire 1 and the variable was categorized as: non-smoker, occasional, or daily smoker. Alcohol intake during pregnancy was analyzed as a dichotomous variable (yes/no). Maternal educational level was divided into three categories . Total energy intake (kilojoules) was collected from the FFQ, and analyzed as a continuous variable. We collected information from Questionnaire 1 about income and the data were divided into three annual income categories: both partners’ <300,000 Norwegian Krone (NOK), participant’s or partner’s ≥300,000 NOK per year, and both partners’ ≥300,000 NOK. Physical activity level is known to be correlated to dietary patterns and obviously affects the net energy intake [–Nine covariates were chosen a priori because of known associations with birth weight: maternal age at delivery, maternal pre-pregnant body mass index (BMI), height, parity, smoking during pregnancy, alcohol intake, educational level, total energy intake, and household income. Data on maternal age was collected from the MBRN. Maternal age was used as a continuous variable. BMI was calculated from pre-pregnancy weight and height as reported in Questionnaire 1, and data were analyzed as categories  <0.05 were considered statistically significant. Linear models were produced in R 3.3.0 [2). Further subanalyses were performed with additional adjustment for chronic diseases, eating disorders, and physical activity during pregnancy.PASW Statistics software version 19 for Windows  was used for statistical analyses and  R 3.3.0 . The fouEnergy and nutrient intakes in the dietary groups reflected the quality and quantity of the food intakes Table . Mean daAfter ranking the pattern scores into tertiles and creating the four non-overlapping dietary groups as described above, the means and SDs for daily food intakes in the four groups clearly reflected the food items typical of the three original patterns  of the babies were SGA and 2446 (3.7%) were LGA. According to the population-based definition, the corresponding respective figures were 5983 (9.1%) and 5797 (8.8%), while they were 9497 (14.4%) and 4485 (6.8%), respectively, according to the customized definition.Women in the high prudent group gave birth to infants with the lowest median birth weight percentile, unadjusted continuous data, according to all three growth standards. Infants born to women in the high traditional group had the highest median birth weight percentile according to all three growth standards, while the high Western and mixed group had quite similar birth weight distribution Fig. .Fig. 1Boz-scores, while the high traditional group was significantly associated with higher birth weight z-scores for all three growth standards .In an adjusted linear regression, with the high Western group as reference, the high prudent group was significantly associated with lower birth weight ds Table . There wIn agreement with the findings for the continuous birth weight data, women in the high prudent group had an increased risk of SGA and decreased risk of LGA Table . This tr2 but not in the overweight group . Likewise, additional adjustment for maternal recreational physical activity did not influence the results (data not shown).n = 62,415 mother–baby pairs) and found that the associations between the dietary groups and SGA were unchanged (data not shown).Finally, we repeated the analyses for term deliveries only  with seven different dietary patterns, no associations were reported between any of the patterns and SGA or LGA [In contrast with our findings, in a Danish cohort of almost 45,000 women studied with a data-driven approach, Knudsen et al.  found thA or LGA .The comparison of different studies is complicated by strong correlations between nutritional habits, culture, and tradition. In addition to lifestyle factors, genetic factors strongly affect birth weight and lower comparability among different populations . DiffereThe major strength of this study is the large number of women, representing all parts of Norway and all socioeconomic groups. Another important strength is the prospective design; that is, dietary intake was assessed before the outcome was known. Using different definitions of SGA and LGA makes the results more comparable to those from other parts of the world. We also have information on a number of covariates, many of which have also been validated in other studies and shown to be accurate measures, for example, smoking . DespiteThe prudent dietary pattern was associated with lower birth weight and the traditional dietary pattern was associated with higher birth weight in comparison with high adherence to the high Western diet. High adherence to the prudent pattern was associated with increased risk of SGA and reduced risk of LGA. Conversely, high adherence to the traditional dietary pattern was associated with lower risk of SGA and higher risk of LGA. Our results suggest that, in addition to food quantity, food quality may have an effect on birth weight.Supplementary Table S1


Polypropylene has been used in radio-frequency capacitors and has also started to be employed in cables as insulation. The objective of this study was to evaluate the electrical properties of polypropylene filled with natural clay as a nano-material. Polypropylene samples having 0%, 2% and 6% natural clay by weight were exposed to 60-Hz sinusoidal voltages at two different rates of rise. The breakdown voltage of each sample was recorded at these different ramp rates. Also, the Root-mean-squared (rms) current was measured as the voltage was increased across the test samples. The important findings of this study were (a) the breakdown strength of the natural nanoclay-filled polypropylene was higher than the unfilled polypropylene, and the optimum concentration of nanoclay appeared to be 2% by weight; (b) the current density as a function of the electric-field intensity indicated a non-linear behavior with saturation, and the saturation onset took place at a higher electric-field intensity in nanoclay-filled polypropylene, wherein 2% nanoclay seemed to be the optimum concentration as well for the onset electric field of saturation. Nano-particles have found a wide variety of applications in order to modify the properties of polymer dielectrics. Since the beginning of this millennium, studies into nanocomposites with polymer-based materials for electrical applications as dielectrics and insulators have been intensified considerably in order to better understand various phenomena in these materials so that their effective usage will be possible in numerous applications .It is known that the addition of nano-particles to a polymer seems to enhance the overall properties of the composite . In suchA polymeric nanocomposite based on PE filled with nanometric size metal-oxide has been introduced recently due to the lower charge mobility than that of pure PE as a potential candidate for HVDC cable insulation . Also, aAlong with PE and other polymers, polypropylene (PP) has found a place in electrical systems as a dielectric . The isoDifferent electrical properties of PP have been investigated in various studies ,15,16,17l.67 Mg0.33 (Na0.33)]Si4 Ol0 (OH)2. Since PP is a hydrophobic material and purified natural clay is hydrophilic, they are immiscible thermodynamically [®20A [®3150 with a composition of PP-maleic anhydride modifier (PP-MA) [®B 225, an antioxidant [The polyolefin used in this study is isotactic PP, and the natural nanoclay is montmorillonite, [Alamically . Consequamically . After tly [®20A ,23, the  (PP-MA) . Also inEach nanocomposite sample had a thickness of 135 μm with a variation of ±10%. Before being situated between two stainless-steel electrodes, the thickness of each sample was recorded using a micrometer. Each electrode had a circular flat surface with a diameter of 33 mm and rounded edges in order to minimize electric-field non-uniformity. The electrode-film assembly was immersed into highly refined transformer oil in a stainless-steel test cell.Because the thickness of the test sample is much smaller than the diameter of the electrodes, the electric-field distribution in the test sample can be considered as almost uniform. Also, in order to avoid the possibility of breakdown in air around the electrode edges, the assembly of the electrodes and the test sample were immersed in highly refined transformer oil filling the test cell.The rating of the high-voltage test transformer was 220/100,000 V (rms) with output power of 5 kVA at 60-Hz.A capacitive voltage divider CM from Messwandler Bau, Germany, was incorporated in order to measure the applied rms voltage to the test sample as indicated in Eb is the breakdown strength, Vb is the breakdown voltage and d is the thickness of the sample. Since the electric field has almost a uniform distribution in the samples, the breakdown strength can be calculated as:As can be seen from In Experimental procedures were the same when a ramp speed of 1050 V/s was used as with the 90 V/s rate of rise in order to assess the breakdown strength of the test samples. At this ramp speed, the breakdown strength of 5 samples is also presented in At a higher ramp-speed of 1050 V/s, nanocomposite with 2 wt % nanoclay content indicated about 3.9% improvement on the breakdown strength as compared to control PP that has no natural clay content. The increase in breakdown strength, however, was not, once again, as substantial when the nanoclay content was increased from 2 to 6 wt %. When compared, the variations of the breakdown strengths as a function of the natural clay concentration in PP at two different rates of rise of the applied voltage, in The current-density vs. electric-field intensity characteristics of PP and its nanocomposites with natural clay were also evaluated within the scope of this study. In this context, current density essentially corresponds to the applied current injected to the sample. Due to the nature of dielectrics, applied current can be modelled with two components under sinusoidal steady-state. These currents are the conduction current due to the transport of charge, and the displacement current due to the rate-of-the-change of electric-flux density with respect to time in the material. Using a Tektronix oscilloscope  the current leads the applied voltage by 1.5359 rad at 60-Hz. Because this angle is close to π/2 rad, which is for perfect dielectric, it is apparent that the conduction current is insignificant as compared to the displacement current. Thus, the conduction current at 60-Hz is neglected and the measured current is considered to be approximately the displacement current. In fact, the amount of error in the current introduced with this approximation was calculated ±0.061%. Therefore, the current density is primarily governed by the permittivity of the material under consideration.J is the current density (A/m2), I is the applied current to the material and A is the surface area of the material effectively sandwiched between the electrodes (m2). J vs. E characteristic is non-linear, with saturation for the control PP.Here, ω is the angular frequency (rad/s) and D is the electric flux density vector (C/m2).From Maxwell’s equations, the displacement current density is directly proportional to the electric-flux density in frequency domain as:Also, due to the constitutive properties of dielectric materials, electric flux density is directly related to the permittivity of the medium along with the applied electric-field intensity as:E is the applied electric-field intensity (V/m) and ε is the permittivity of the medium (F/m), which can also be expressed as:εo = 8.85 × 10−12 F/m being the free-space permittivity and Here, material . If the Relative permittivity, which is often referred to as the dielectric constant, is also defined as:For the unloaded PP, raising the electric-field intensity to 30 kV/mm from zero yields a permittivity of 2.27, which is nearly constant, as shown in J vs. E characteristic as stated in standards [It should also be noted that these values may differ from one kind of PP to another since the composition may have different concentrations of PP along with different component materials. Nevertheless, the relative permittivity of PP under consideration is within the limits of the dielectric constant (2.2–2.5) in the linear region of the tandards .The variation of current density with the electric-field intensity is given in J–E characteristic of the material. For electric fields higher than 35 kV/mm, however, there is a substantial reduction on the relative permittivity from 2.25 to 1.38 at 110 kV/mm due to the non-linearity of the material.It can be observed from J vs. E characteristic as pointed out in PP filled with 6 wt % natural clay also revealed a non-linear, saturation-type The relative permittivity of this nanocomposite appears to be nearly constant at 2.24 up to the electric-field intensity of 33 kV/mm as presented in Similar to the plain PP and the nanocomposite filled with 2 wt % nanoclay, the material exhibits a reduction in its relative permittivity to 1.42 at 110 kV/mm. As can be observed from these results, natural nanoclay in PP did not contribute significantly to the current density vs. electric-field characteristic or to the relative permittivity.The dielectric breakdown of PP is affected by the polymer morphology when PP is subjected to sinusoidal voltage. PP usually breaks down electrically in locations where the material is less dense. Andritsch et al. indicated that the smallest breakdown voltages took place in the interspherulitic zones of PP, whereas the highest breakdown voltages took place in the spherulites regions . TherefoAlso, an optimum content of nanoclay is proposed according to the results on the breakdown strengths in this investigation. It was observed that the breakdown strength did not improve substantially for the nanocomposite filled with 6 wt % nanoclay when compared to 2 wt %. Therefore, nanocomposite loaded with 2 wt % natural clay was considered to be the optimum amount of natural clay. Abou-Dakka et al. concluded similarly with the similar nanocomposites of PP in which all the homocharges could possibly be reduced for the optimum nanoclay content .When the rate of rise of the applied voltage was increased from 90 to 1050 V/s across unfilled PP and its nanocomposite samples, a significant increase was observed on the breakdown strength of the materials. The increase in breakdown strength is due to the insufficient time to transfer energy to the particles in the unfilled PP as well as in the natural nanoclay-filled samples to initiate the ionization activity when the rate of rise of the applied voltage is higher. Thus, a higher electric-field intensity is required to maintain higher energy for the inception of ionization activity that might possibly cause a sustained discharge leading to the failure of the material.J–E characteristics of the materials, the relative permittivity remains almost constant since the net polarization of the material takes place at a constant rate of change with respect to the electric field. But, with respect to a higher electric field, net enhancement rate of polarization diminishes and leads to a significant reduction in the permittivity of the material. In a another work it was also stated that although polymeric materials have almost constant permittivity at relatively low electric fields, at high electric fields, substantial drops could be seen in the permittivity [The saturation effect in the current density vs. electric-field intensity characteristics in unfilled PP and its nanocomposites loaded with natural nanoclay can be explained based on the net polarization of the materials. From Equation (3) it should be noted that the electric flux density is an angular frequency reduced displacement current under sinusoidal steady state. Also, the electric flux density is proportional to the net polarization of the material as indicated in Equations (4)–(6). When the applied electric field is raised starting from zero up to a certain level, the displacement current density increases linearly since it is directly proportional to the net polarization in the material. Above that certain electric-field intensity, on the other hand, the rate-of-change of net polarization with respect to the electric-field intensity becomes lower. Consequently, the saturation effect starts occurring in the material. In the linear region of the ittivity .As discussed in Current density vs. electric-field intensity characteristics of all the materials in this study have shown very similar trends with some saturation effect. When the electric field was raised in the material up to a certain level, it seemed to have a linear behavior. However, above a critical applied electric field, materials responded non-linearly as discussed in Surface modification seems to be important for the contribution of the filler material to the properties of nanocomposites as indicated in ,13,14,27


The cohort was 46% female; mean birth weight and gestational age (GA) were 998 g and 27.3 weeks. Multilevel linear growth models assessed changes in growth z-scores short-term (to NICU discharge) and long-term (post-discharge), controlling for amount of DM or formula received in first 28 days of life, NICU length of stay (LOS), birth GA, and sex. Z-scores for weight and length decreased during hospitalization but increased for all parameters including head circumference post-discharge. Short-term growth was positively associated with LOS and birth GA. A higher preterm formula proportion, but not DM proportion, was associated with slower rates of decline in short-term growth trajectories, but feeding type was unrelated to long-term growth. In conclusion, controlling for total human milk fed, DM did not affect short- or long-term growth.Mother’s own milk (MOM) reduces the risk of morbidities in very low birth weight (VLBW) infants. When MOM is unavailable, donor breastmilk (DM) is used, with unclear impact on short- and long-term growth. This retrospective analysis compared anthropometric data at six time points from birth to 20–24 months corrected age in VLBW infants who received MOM supplements of preterm formula ( Mother’s own milk (MOM) is the recommended form of nutrition for all infants, particularly for those born preterm, in whom MOM reduces the risk of neonatal morbidities, including necrotizing enterocolitis (NEC), and improves long-term neurodevelopmental outcomes ,2,3. Witz-scores in very low birth weight  infants at six time points: Birth, 32 weeks corrected gestational age (CGA), NICU discharge, and 4, 8, and 20–24 months corrected age (CA) at NICU clinic follow-up visits. We hypothesized that DM-fed infants would have persistently decreased anthropometric z-scores post-discharge in comparison to infants supplemented with formula.DM has been associated with decreased rates of short-term, in-hospital growth compared to both preterm formula and MOM ,11,12,13The sample included 321 VLBW infants admitted to an urban level III NICU in Chicago in two time periods: A “pre-DM” group using preterm infant formula as a supplement to MOM if MOM was unavailable/insufficient in quantity, and a “DM” group that supplemented MOM with pasteurized DM instead of formula. Data from two cohorts of infants were combined for this comparison: 160 infants admitted between January 2011 and December 2012 who were part of a U.S. National Institute of Health-funded prospective cohort study, for whom inclusion and exclusion criteria have been detailed previously , comprisThis secondary analysis of cohort data compared growth outcomes for the pre-DM and DM groups. In both groups, parenteral nutrition began on the day of birth, with unchanged composition and protocols between the pre- and DM groups. Parental nutrition was weaned as feedings were advanced until discontinuation when feedings reached 120mL/kg/day. In the pre-DM group, enteral feedings were started as soon as MOM was available, a practice that may have delayed feeding initiation for up to 5 days post-birth. If MOM was unavailable by that time, 20 kcal/oz  preterm formula was begun. In the DM group, enteral feedings were also initiated as soon as MOM was available, but by day 2 post-birth, DM consent was sought if MOM was unavailable in sufficient quantities. Pasteurized DM from a milk bank in Indiana  was used to supplement insufficient MOM through 35 weeks corrected GA (CGA), following a gradual wean to preterm formula starting at 34 weeks CGA. Preterm formula was used if families did not provide DM consent. In both groups, formula was changed from 20 kcal/oz  to standard 24 kcal/oz  preterm formula once feedings reached 140 mL/kg/day and caloric density was further increased if needed to maintain adequate growth, at the clinical team’s discretion. Bovine-based human milk (HM) fortifier , was added to both MOM and DM when feeding volumes reached 140 mL/kg/day to approximate 24 kcal/oz , assuming a baseline of 20 kcal/oz . Macronutrient analysis of MOM or DM was not performed. Additional modular protein (0.5–1 g/kg/day) was routinely provided to all infants receiving DM but was added to MOM only if needed for poor growth and/or low blood urea nitrogen (<9 mg/dL) . StandarData were abstracted from the primary study and electronic medical records (EMR). Infant demographics (length of stay (LOS), GA at birth, birth weight, sex, race/ethnicity) and daily enteral intake  were collected for the first 28 days of life (DOL) during the NICU hospitalization. The proportion of each type of enteral intake was calculated for each infant as the cumulative volume of MOM, DM, or preterm formula received divided by total volume of enteral feeds (MOM + DM + formula) received over the study time period. Weight (g), length (cm), and HC (cm) were recorded at birth, at 32 weeks CGA, at NICU discharge, and at NICU clinic follow-up visits at 4, 8, and 20–24 months CA. Parent-reported post-discharge liquid diet  was collected from the EMR from NICU follow-up clinic visits.z-scores using Olsen growth charts for premature infants through 40 weeks CGA and World Health Organization (WHO) growth charts thereafter [z-scores. Small for gestational age (SGA) was defined as birth weight below the 10th percentile [z-scores as a function of overall time in months (centered at NICU discharge) and a post-NICU time parameter that had values of zero for all time until NICU discharge, followed by number of months since discharge. The combination of these two parameters afforded the estimation of two growth estimates: (1) Short-term, in-NICU growth  and (2) long-term, post-NICU growth . The post-NICU time parameter represents the deviation from the in-NICU growth trajectory that occurs after discharge. Models controlled for NICU LOS, GA, sex, group (pre-DM vs. DM), formula proportion, and DM proportion. NICU LOS, GA, sex, and group were mean-centered; formula proportion and DM proportion were not centered. Thus, the overall intercept (z-score at NICU discharge) and growth (change in z-score over time) parameters represented model estimates for “average” infants  who received 100% MOM  and thus constituted the reference group. Interaction terms with overall time and the time parameter for all covariates allowed estimates of the effect of the covariates on short-term and long-term growth. To ease interpretation of the model parameters, planned estimates were calculated for short-term and long-term overall trends and for the effects of each covariate on both short-term and long-term growth. All models employed maximum likelihood estimation, as is standard for longitudinal growth models. Under the assumption of missing at random, this allowed unbiased growth trajectories to be estimated for each infant, regardless of missing measurements. For the growth models, we employed the Benjamini–Hochberg approach to control the false discovery rate associated with testing multiple hypotheses [Infants’ weight, length, and HC measures were converted to ereafter ,20. Corrrcentile . All varrcentile  allowed potheses . SPSS vep < 0.05 for all.Characteristics of the infants and feedings are summarized in Anthropometrics are reported for both groups from birth to 20–24 months CA in Z-scores declined significantly for weight and length during NICU hospitalization from birth to discharge for the cohort as a whole (z-scores that compare estimates for the reference (100% MOM) infants to estimates for infants at the 95th percentile of formula use for the cohort (which corresponded to receiving 94% formula) or the 95th percentile of DM use (which corresponded to receiving 95% DM). a whole . Longer z-scores increased significantly for all parameters  of infants were still receiving MOM at the 4- and 8-month CA follow-up visits  than in the pre-DM group (0.85 MOM), despite a unit policy of a two-day “waiting period” prior to consenting for DM to encourage maternal pumping and MOM use . In NICUOur study is one of the first to examine the long-term, post-NICU growth implications of fortified DM to supplement an enteral diet with a high proportion of MOM in the critical first month of life. A strength of our study is the racially and ethnically diverse urban cohort of VLBW infants with detailed daily feeding data for the first month of life, as well as multiple growth data points during and after the NICU stay through to 20–24 months CA. However, although we had a relatively large sample size of 321 infants, it was a single center study with a strong MOM-feeding program, so the results may not be as applicable to NICUs with lower MOM and higher DM use in early life . A limitOur institutional practice was to discharge infants on unfortified MOM and/or transitional 22 kcal/oz  formula, which may differ from other institutions that continue MOM fortification post-discharge. Since published and unpublished studies have not consistently demonstrated beneficial effects of post-discharge fortification on long-term growth in very preterm infants, it is undetermined if our results would have been impacted by different post-discharge fortification practices ,47,48. Az-scores declined during the NICU hospitalization, with no relationship detected as a function of DM proportion, but with a positive association between growth z-score and formula proportion. However, after discharge, the overall growth pattern reversed from decreasing to increasing z-scores without variation by in-NICU feeding type. These findings support the use of DM early in the post-birth period to mitigate potential risk from formula without compromising long-term growth. Although fortified DM in this cohort did not appear to negatively impact growth through to the age of two years, additional studies are needed to delineate longer-term implications of DM feeding in VLBW infants.VLBW infant weight and length