The present work is devoted to assessing the influence of discreteness of electric charge distribution in the double electric layer on the characteristics of the electric fields and their force action in capacitor structures with small interelectrode gaps. Due to the fact that modern technologies often use submicron-sized interelectrode gaps, it is no longer possible to consider the electrodes uniformly charged because of the discreteness of the electric charge. The corresponding development of a mathematical and physical model for the study of a non-uniform electric field is suggested. Numerical calculations are carried out, expressions, criteria, and results that are convenient for practical evaluations are obtained. The physical and mathematical model for force characteristics of a non-uniform electric field is developed. With a sufficiently small size of the interelectrode gap, the integral force effect of discretely distributed charges can be significantly higher than with a uniform distribution of the same charge. At reasonable surface charge densities, these phenomena are usually observed at interelectrode gaps less than tenths of a micrometer. The principle of their operation, one way or another, is based on the use of an electric field (EF). These include:-Extremely low power consumption. If an electret is used as a source of electrical energy, then the device may not consume electrical energy at all during operation;-Constructive simplicity and relative cheapness of the corresponding devices. An example of such devices is electret microphones, the production of which in the world now exceeds 90% of all microphones produced;-High technical characteristics of electrostatic devices. A lightweight, low-inertia membrane is often used as one of the capacitor plates. This makes it possible to transform signals with minimal distortion, to obtain high-speed CS;-The prospects for use in MEMS technology . This isThe interest in technologies and devices based on EF is due to a number of their advantages, the main of which are:One of the possible practical applications of the considered influence of the EF is the possibility of increasing the adhesion of coatings deposited to substrates, both after and during their deposition, as well as a number of other effects ,20,21,22In  a theory(a)Mutual diffusion of materials to be joined. An increase in the adhesion strength of the surfaces to be joined can be ensured by the diffusion of atoms and molecules, which leads to strong adhesion of the materials to be joined. The sharp interface of the contact in the process of diffusion turns into a transition layer. This process is the basis of the known diffusion bonding method and is possible using external pressure and elevated temperatures ,3.(b)7 Pa), and to achieve not only temporary, but also permanent connection of parts. This method of joining is called electroadhesive, and the corresponding technology is called anodic fit or anodic bonding [Formation of a layer of electric charges of the opposite sign on the contact. The role of this factor in the phenomena of adhesion, in our opinion, is often underestimated. Indeed, as shown in ,7, an in bonding .Derjagin et al in  point ouAt present, intensive development of theoretical concepts of adhesion continues, while the authors note the great influence of the electrical component of adhesion ,20,21,22A feature of electrostatic forces is that they become very significant only at small distances between the electrodes. If we are talking about DEL, then its thickness is usually of the nanosized dimensions. An increase in the electric field strength in DEL due to the use of an external voltage source leads to the appearance of additional charges at its boundaries. In this case, as will be shown below, the distance between these additional charges (electrons) can be not only comparable, but also can be much greater than the DEL thickness. With this state of affairs, the concept of “uniformly charged plane” obviously loses its meaning, since charges on the plane will be distributed discretely . It is cThus, the examples considered show that the use of interelectrode gaps of submicron thickness is currently in demand. At the same time, the features of the behavior of EFs in such gaps have not been sufficiently studied. Therefore, the purpose of this work is to develop a physical and mathematical model for calculating the EFs and their force action when using interelectrode gaps of submicron size, taking into account the discreteness of the charge distribution. The quantitative results that will be obtained in this case will make it possible to reveal the limits of applicability of the classical approach to solving this problem and to reveal the quantitative features of EFs at small interelectrode gaps.Let us consider in more detail the effect of the discreteness of the distribution of electric charges on the force action of the electric field. The need for such a consideration was undertaken because of the fact that when obtaining electroadhesive joints, such small interelectrode gaps are used in them that it is no longer possible to consider the electric field to be uniform because of the discreteness of the electric charge. In addition, as shown by indirect experiments related to determining the strength (work of separation) of electroadhesive joints, depending on the various parameters of the technological process, the classical model based on the concept of “uniformly charged plane” gives underestimated values of electroadhesive forces. When developing a more accurate model, we will take into account that the length of the wave function of the electron itself in the material usually corresponds to not more than the units of nanometers, i.e., electrons can be roughly considered point charges.p is the ponderomotive pressure, e = 1.6 × 10−19 C) and, therefore, the electric field of point charges system is not homogeneous. At the same time, the model for the calculation of the ponderomotive effect of the electric field on the basis of Equation (1) is acceptable in the case when the distance between the planes of the DEL is much greater than the distances between the discretely spaced elementary charges. In the case when the distance between planes of DEL is commensurate with the distance between the discretely spaced elementary charges, e.g., during the formation of electric adhesion compounds, the heterogeneity of the electric field manifests itself [The formula of a plane capacitor is usually used to calculate the interaction force caused by the attraction of opposite charges, distributed on the planes of the DEL:s itself ,6. Thus,−4 C/m2. As it is known that the electric field strength created in a vacuum  will not adequately reflect the real physical situation. As an example let us consider a uniformly charged plane with surface charge density 6 V/m.Calculations according to Equation (2) give the value 2 area that obviously corresponds to the specified value of electric field strength isIt is precisely such and large in order of magnitude of the electric field strength that are used to obtain electroadhesive joints. In this way, the amount of elementary charges per 1 m14 m−2. Consequently, the average distance between them isCalculations according to Equation (3) give the value −8 m.Calculations according to Equation (4) give the value a = 0.08 μm, and for treated by 14th class—Ra = 0.01 μm.Thus, at the distance between the DEL planes corresponding to the order of tenths of a micron , the Equation (1) can give untrue value. Note that it is the abovementioned distances between the charges that are used in the process of obtaining electroadhesive joints, since the gap between the parts to be joined is mainly due to the roughness of the surfaces to be joined, which are usually processed no worse than in the 11th class of cleanliness. Moreover, in the process of obtaining electric adhesion compounds surfaces are closed together and the distance between them decreases. For reference, we note that the value of the arithmetic deviation of the rough surface profile for items treated by the 11th class is RThus, at small distances between the charged planes the discreteness of electric charge distribution should be taken into consideration. Indeed, in experiments ,6,7 pondIt is known that with a continuous uniform distribution of electric charge over the plane, the strength of the electric field created isc”: For definiteness, let us add the index “This is an infinite plane with discontinuous distribution of charges and with the same as in Equation (5) area-averaged surface charge densitys, etc., ,22,23,24We define the field at a point M lying on the dotted line. Point M is chosen to consider the plane lying opposite one of the charges, because the charges of opposite signs of the DEL are located not chaotically, but correlated—against each other. To calculate the strength of the electric field The representation of Equation (7) in the form of a double series is due to the symmetry of the distribution of charges in each of the quarters of the plane relative to the charge onto which the point M is projected, the latter is represented in the formula by a separate term . The sumcrr, at which the values of crr ≥ a, which with the actually used values of σ corresponds to submicron dimensions. At the same time, as r approaches zero, the value of r > 10−9 m. At distances much smaller, the complex nature of the force interatomic interaction begins to manifest itself—to the point that at very small distances the atoms will repel and our calculation becomes meaningless.A comparison of the obtained values of the electric field strengths at different charge densities is graphically presented in l⋅a, where l is the number of planes, and a is the minimum distance between charges.Consider the question of the location of elementary charges in a layer of finite thickness, consisting of several planes . To objem and n gives the total field from one plane, summation over l—from L planes.Let’s calculate the strength of the electrostatic field in the system shown in L = 1) and therefore for the electric field strength created by discrete charges the expression Equation (2) with a not very large number of layers and not too close a distance to the observation point can be used correctly.Numerical calculations show the insignificance of the influence that the layer thickness has on the resulting electric field. Therefore, it is enough to consider the case of only one layer (n (it is important to note that the order of the sums over n and over m is not important here because of the continuity of the integrand). Numerical calculations have shown that series Equations (7) and (8) converge slowly, i.e., the maximum values of n and m should be taken large enough. In this case, the contribution of each subsequent term with an increase in n and m is very small compared to the current value of the sum Equations (7) or (8). This gives reason to replace the summation by integration. The corresponding integral was analyzed when extending the domain of the function from the set of integers to a continuous positive semiaxis. The result of the corresponding calculations is a simple function that allows you to accurately calculate the fields of discretely distributed charges:r is the distance from the layer containing discretely distributed charges to the point at which the electric field strength is determined.Direct use of Equations (7) or (8) leads to the computational problems, therefore, for the practical use of the obtained expression, it must be approximated by an analytical formula—which would deviate least from the value of the sum of the series under consideration. Let us start by considering the sum over the variable a is the distance between the charges. From what we have obtained, it can be seen that as Let us analyze the resulting expression Equation (9). First consider the situation when the charges are very dense, i.e., k depends on the area-averaged surface charge density σ and the distance Thus, based on Equation (9), we conclude that to take into account the discreteness of the charge distribution, we can use the simple formulaBelow is −6 C/m2 and −9 m differs from the traditionally calculated value by 400 times. In this regard, we recall that our calculation was carried out for the point M, which lies strictly opposite one of the charges that create the electric field. Moreover, for this situation, the calculation is fair. It is the presence of such large local electric fields and, accordingly, local forces that, in our opinion, explains, in particular, the abovementioned ability of the double electric layer to neutralize the embryonic cracks, formed in the plane of adhesive contact and subject to expansion under external influence. The calculated value of the electric field strength should not be confused with a certain average field of the considered plane with discrete charges. Indeed, based on Equation (6), it is easy to calculate that in this example, the distance between the charges will be a = 4 × 10−7 m. Thus, the height of the considered point M above the plane, x and y. This is a rather difficult task and is not considered in this work. Therefore, it should be emphasized that if we understand by The data in x and y) will not be decisive. In this regard, it has already been noted that series Equation (7) converges slowly, which physically means that in this case the field strength is determined mainly by the cumulative effect of the set of charges on the plane, and not by the specific nearest charge. Thus, at In this case, the position of the point M above the plane , it is assumed that the surface charge density σ can be of any nature, for example, chemical. When making electroadhesive connections, charges are created by using an external electrical voltage. In this case, charges of opposite signs, created by an external source, can be considered as charges of a flat capacitor and therefore can be written:Expressing −9 m) and submicron (−7 m) dimensions of the gap shows the values of the electric field strengths a calculated on the basis of Equation (6) are given.We emphasize once again that for Considering that the force (ponderomotive) action of the electric field is proportional to the square of the electric field strength, the effect of the discreteness of the charge distribution can in this sense be very significant even with a relatively small value of the charge itself. The following simplified calculation illustrates this idea. Let us consider the situation when the charges on the electrodes are far enough from each other and therefore we can assume that the interaction occurs only in pairs between charges opposite to the electrodes, a2. Then the corresponding pressure Let us calculate for this case the ponderomotive pressure arising between the electrodes and causing their attraction. To do this, we take into account the Coulomb’s law, as well as the fact that for each pair of attracting charges there is an area The ponderomotive pressure K. From Equations (14) and (15) we obtain:Let us find out how significant an increase in pressure can be due to the discreteness of the charge distribution with a constant value of the total number of charges on the electrodes. For this, consider the ratio of From Equation (16) it is seen that for sufficiently small r, i.e., when the electrodes are located close, we get The obtained values correspond to the nano- and micro-dimensions of the interelectrode gap.Let us quantify the indicated effect of pressure increase. For this, we calculate the coefficient −4 C/m2 typical for the use of an electric field , corresponds to a field strength of ≈107 V/m), and the thickness of the interelectrode gap will be −8 m. In this way, as can be seen from From what has been considered, it is clear that the calculation must be performed at values Calculations show that the effect under consideration for small sizes of interelectrode gaps (in comparison with the distance between charges on them) can be very significant and lead to an increase in pressure by orders of magnitude. It should be noted, however, that at nanometer sizes of interelectrode gaps, various quantum phenomena (primarily tunneling) can be observed, which can, to a certain extent, complicate the manifestation of the above effect. In particular, as the size of the interelectrode gap tends to be zero, the length of the electron wave function may be comparable with the size of the gap. Obviously, in this case, you cannot use the Coulomb law.Direct experimental verification on the equipment available to us turned out to be difficult. Therefore, we used a large amount of our experimental results accumulated in the study of electroadhesive joints. From the most general concepts of adhesion, it follows that it is possible to regulate the adhesion by applying an electric voltage to the objects to be connected. Despite the fact that this possibility has been proven experimentally, it seems that currently the corresponding techniques are not widely used.In the formation of electroadhesive joints, as already noted, the force action of an electric field in micro- and nanoscale interelectrode gaps is used. It is possible to achieve large values of electric field strengths, in particular, because of the migration polarization of the dielectric, “glued” to the metal or semiconductor. In this case, the charge is accumulated in a thin dielectric layer near the anode.The scheme for manufacturing electroadhesive joints is shown in In a number of our works ,6 both tTo quantify adhesion, the scratch method is often used ,24. Howem is the mass of the load, g is the acceleration of gravity, b is the width of the film being torn off, In this case, the work of tearing off the film (work of separation) is found by the expression (20)A=mgWith high adhesion of the film to the substrate, only partial peeling of the film from the substrate occurred during measurements. In this case, according to the results of visual observations, the relative fraction of the exfoliated area of the film Of course, the proposed approach is not rigorous and was used only because it makes it relatively easy to get a quantitative idea of the adhesion of the film to the substrate, when it exceeds the adhesion of the scotch tape to the film in a part of the area or the strength of the thin foil used as a coating. There are other methods for determining adhesion, in particular, various modifications of the scratching method are widespread ,24. ThesThe scope of this work does not allow us to consider this model in detail. The presentation of this technique is the subject of a separate article. Therefore, we will restrict ourselves here to the most simplified, but containing the basic ideas about the issue under consideration. We point out that the relative fraction of the actual contact area, obviously proportional to the strength of the resulting joints  for some time , a charge accumulates in the region We have carried out numerous experiments to measure the strength (work of separation) of various electroadhesive “glues” depending on the various technological factors—the temperature of the joint, the roughness of the surfaces to be joined, the holding time, etc. All of them confirm the decisive influence of the electric field on the strength of the joints.Obviously, the most direct factor affecting the magnitude of the electric field is the electric voltage, zR = 0.04 μm—the roughness of the surfaces to be joined), as shown above, can turn out to be significant. When plotting curve 2, It was noticed that at low voltage values and, consequently, low surface charge densities created by it, the calculated dependence , r, as above, we mean the distance between the interacting charges. In our case, these are charges on the aluminum foil and in the region zR = 0.04 μm, to the minimum, corresponding to interatomic distances. Thus, an accurate calculation is rather difficult, so we will restrict ourselves to only rough estimates. In this regard, r from the specified range.Under K cannot be less than one.Some cells of Curve (2) in  contact , in EquaK was chosen for the case r = 4 × 10−9 m , will be of the order of 1 kPa. The adhesion strength of various joints can be at the level of the cohesive strength of the solids being joined, and the latter is of the order of 100 MPa. Thus, this approach does not show a significant contribution of the electrostatic component to the adhesive strength of the joint. At the same time, in the above literature data, there are numerous evidences of a significant influence and various noticeable manifestations of the electrostatic component of adhesion.With the help of the approach developed in the work, an explanation can be given for some experimental data, which are difficult to explain with the classical approach. For example, as a result of classical calculations when analyzing electric fields in most real situations, the values of electric field strengths are obtained, as a rule, no more than 10In our opinion, this situation can be easily explained using the approach developed in this work: the local values of the field strengths can be significantly higher than those indicated above, and for small interelectrode gaps, the force action of the electric field can significantly increase in comparison with the case when the discreteness of the charge is not taken into account. The considered features of the electric fields can, in particular, explain the strong influence of the electric field on the adhesion of conducting films deposited on dielectric substrates.Mathematical equations provided in this work are believed to be essential for calculating the various effects in MEMS—technologies where small distances and the electrostatic principle of controlling are often used. Besides, the developed method can be used at calculation of the electric effects resulting from not uniform distribution of a charge caused, for example, by structure defects.It should also be noted that the approach developed in this work for calculating electric fields can also be used for the case of relatively large interelectrode gaps, when the source of the field is macroscopic regions unevenly charged over the surface. An example of such objects are the so-called matrix electrets .The performed calculations have shown that at small distances to the charged plane, electric fields may appear that significantly exceed the values calculated by the classical formulas. The considered field with a discrete charge distribution is inhomogeneous; therefore, due to gradient forces, it can have a pulling force effect even on electrically neutral objects, for example, atoms deposited onto a substrate, and have other manifestations that can find practical applications. The performed calculations show that the considered effects become significant, as a rule, at submicron sizes of the interelectrode gaps. The expressions obtained in this work make it possible to quantitatively evaluate the influence of the discreteness of the charge distribution on the features of electric fields in small interelectrode gaps. In this case, the term “charge” does not necessarily mean the charge of an electron. These can be regions of relatively large dimensions on the surface of the material, in which, for one reason or another, there is an increased density of electric charge.


The universal two-child policy in China which was announced in October 2015 might affect the caesarean birth rate. Few studies reported the caesarean birth rate affected by the policy especially after the universal two-child policy period. This study aimed to demonstrate the caesarean birth rate changes before and after the universal two-child policy and further elaborate the underlying influencing factors.This nationwide, retrospective, population-based study was based on National Free Pre-Pregnancy Check-ups Project (NFPCP). Couples planning to conceive in next 6 months were encouraged to participate in NFPCP. Baseline information was collected using a standardized questionnaire with a face-to-face interview, and medical examinations were conducted. Two rounds of follow-up interviews were then conducted by trained nurses to update pregnancy status and outcomes. NFPCP participants who completed deliveries before December 31, 2018, were included in the current study. We used marginal effect of year to examine the trend of caesarean birth rate over time and interrupted time series (ITS) analysis to determine impacts of the universal two-child policy on the trend of caesarean birth rate.A total of 9,398,045 participants were included in the final analysis. High-risk factors to increase caesarean birth rate were identified. In the current study, the standardized caesarean birth rate declined from 34.1% in 2013 to 31.8% in 2015 and increased to 35.6% in 2018. ITS analysis showed the caesarean birth rate decreased by 0.1% (95% CI 0.1–0.1) per month before the release of universal two-child policy, 1.3% (95% CI 0.6–2.1%) absolute drop during the policy release month, and increased by 0.2% (95% CI 0.1–0.2%) per month after the policy implementation. For the period after the policy release, the increasing trends were observed in rural participants and urban primiparas. The prevalence of caesarean birth rates within China varied regionally.The decreasing trend of caesarean birth rate was reported after immediate release of the universal two-child policy. An increasing trend of caesarean birth rate was observed 2–3 years after the policy. It reminds us that the caesarean birth rate control is a long-lasting process and all the strategies need to be continually reinforced. Caesarean birth is a surgery with both short- and long-term risks compared to vaginal birth, such as surgical complications, neonatal intensive care unit admission, and higher costs –3. GlobaConcerns have been raised on overuse of caesarean delivery in all parts of the world. In China, the caesarean birth rate increased dramatically during the past three decades –10. In 1Based on the evidence, national and local health commissions and medical professionals have been working on reducing caesarean birth rate especially nonmedically indicated caesarean birth in China. In the past few years, the government has undergone efforts to curtail the rising caesarean rate and promote vaginal deliveries by launching a number of different programs , 12. TheThe relaxation of the one-child policy in November 2013 allows at least the only-child of either parent to have two children, and the introduction of the universal two-child policy in October 2015 allows each married couple to have two children . For priHowever, few studies reported the caesarean birth rate was affected by the policy especially after the universal two-child policy period. The data regarding the caesarean birth rate change after the two-child policy are needed. The current large population-based study aimed to assess the prevalence of caesarean deliveries from 2013 to 2018 in China and to provide the latest information on caesarean birth rate changes after universal two-child policy era and further elaborate the possible reasons.This nationwide, retrospective, population-based study was based on National Free Pre-Pregnancy Check-ups Project (NFPCP). Supported by the National Health Commission and Ministry of Finance of the People’s Republic of China, NFPCP provides preconception care including free health examinations prior to pregnancy, risk assessments, consultations, and two rounds of follow-ups for pregnancy outcomes for reproductive-aged couples who wish to conceive in the near 6 months. The national project was initially set to promote reproductive health of rural resident couples and then gradually expanded to urban resident couples since 2013. Even after the expansion, the national project still mainly focused on the rural areas. Until 31 December 2018, the NFPCP have covered 3245 counties/districts across 31 provinces in mainland China. Detailed design, organization, and implementation of the NFPCP have described elsewhere , delivery date, foetal number, and birth weight.This study was approved by the Institutional Research Review Board at the National Health Commission. Written informed consent was obtained from all NFPCP participants.2 (kg/m2). BMI < 18.5 was considered underweight, BMI ≥ 18.5 and < 24 were considered normal, BMI ≥ 24 and < 28 were considered overweight, and BMI ≥ 28 was considered obese according to the Chinese population standards , higher education level [OR 1.16 (1.15–1.16)], overweight [OR 1.20 (1.20–1.21)] and obese women [OR 1.53 (1.52–1.54)], and complicated with adverse pregnancy outcomes in previous deliveries [OR 1.35 (1.35–1.36)] et al. were independent risk factors of caesarean birth . It seemed there was no baby boom after the universal two-child policy release as expected.The next question is whether the increase of caesarean birth rate is due to the assumed ‘baby boom’ after the universal two-child policy. The relaxation of the one-child policy has been introduced gradually. The couples who were both only-child were permitted to have two children since 2007. The policy then allowed at least the only child of either parent to have two children since November 2013. Only 13.2% eligible couples applied for permission to have a second child by May 2015 [The optimal threshold of caesarean delivery use is hard to determine. It has been suggested caesarean delivery use reaches an optimal threshold ranging from 9 to 16% to improve maternal and neonatal survival . China wOur study provided more lines of evidence regarding the caesarean birth rate before and after ‘universal two-child policy’ and identified several high-risk factors  for caesarean birth rate have been increased. In the current study, there is a noticeable increase of caesarean birth rate which is identified in 2017 and 2018, only 2–3 years after the ‘universal two-child policy’, although a significant but temporary decrease of caesarean birth rate appeared in 2016. It reminds us that the caesarean birth control is still a long process and all the strategies need to be continually reinforced both in primiparas and multiparas, such as promoting trial of labour after caesarean (TOLAC), performing external cephalic version for breech presentation, providing health education for childbearing women, and providing nutritional management during pregnancy to realize optimal maternal weight gain and avoid macrosomia.There are several limitations to our analysis. One limitation is that we could not distinguish the indications for caesarean birth, such as foetal distress, cephalo-pelvic disproportion, previous uterine surgery, malpresentation, and suspected macrosomia –24. SecoThe decreasing trend of caesarean birth rate was reported after immediate release of the universal two-child policy. An increasing trend of caesarean birth rate was observed 2–3 years after the policy. It reminds us that the caesarean birth control is a long-lasting process and all the strategies need to be continually reinforced. Future studies are needed to further demonstrate the larger picture of caesarean birth rate changes influenced by the universal two-child policy in the long term.Additional file 1 : Table. S1. Association between baseline characteristics and the risk of caesarean birth.Additional file 2 : Table. S2. Marginal effects results revealed that the changing amount of caesarean birth rate.Additional file 3 : Table. S3. Interrupted time series model to determine impacts of the universal two-child policy on the trend of caesarean birth.Additional file 4 : Figure S1. Translated of detailed design, organization, and implementation of the NFPCP (described elsewhere in Chinese). [hinese). .


Risks of adverse birth outcomes in England and Wales are relatively low but vary across ethnic groups. We aimed to explore the role of mother’s country of birth on birth outcomes across ethnic groups using a large population-based linked data set.We used a cohort of 4.6 million singleton live births in England and Wales to estimate relative risks of neonatal mortality, infant mortality and preterm birth, and differences in birth weight, comparing infants of UK-born mothers to infants whose mothers were born in their countries or regions of ethnic origin, or elsewhere.The crude neonatal and infant death risks were 2.1 and 3.2 per 1000, respectively, the crude preterm birth risk was 5.6% and the crude mean birth weight was 3.36 kg. Pooling across all ethnic groups, infants of mothers born in their countries or regions of ethnic origin had lower adjusted risks of death and preterm birth, and higher gestational age-adjusted mean birth weights than those of UK-born mothers. White British infants of non-UK-born mothers had slightly lower gestational age-adjusted mean birth weights than White British infants of UK-born mothers . Pakistani infants of Pakistan-born mothers had lower adjusted risks of neonatal death (adjusted risk ratio (aRR) 0.84, 95% CI 0.72 to 0.98), infant death  and preterm birth  than Pakistani infants of UK-born Pakistani mothers. Indian infants of India-born mothers had lower adjusted preterm birth risk  than Indian infants of UK-born Indian mothers. There was no evidence of a difference by mother’s country of birth in risk of birth outcomes among Black infants, except Black Caribbean infants of mothers born in neither the UK nor their region of origin, who had higher neonatal death risks .This study highlights evidence of better birth outcomes among UK-born infants of non-UK-born minority ethnic group mothers, and could inform the design of future interventions to reduce the risks of adverse birth outcomes through improved targeting of at-risk groups. Disparities in birth outcomes across ethnic groups persist in high-income countries, with minority ethnic group infants experiencing poorer birth outcomes than those of majority groups.In the UK, increasing proportions of births occur to women born outside the UK. In 2017, 28.4% of births occurred among women who were born outside the UK,Much of the research on migration and health outcomes focuses on comparing the birth outcomes of migrant women to those of the native women.With the availability of national linked data that includes ethnic groups of babies, mothers’ country of birth and birth outcomes for all babies born in England and Wales since 2005,st January 2006 to 31st December 2012.We analysed data from the cohort of all singleton live births occurring between 22 and 43 weeks of gestation in England and Wales from 1Statutory birth and death registration data for England and Wales were linked to the National Health Service Numbers for Babies (NN4B) birth notifications system by the Office for National Statistics (ONS). The NN4B system included additional characteristics such as ethnic group of babies and gestational age at birth, which were not collected at birth registration. The data linkage and its evaluation have been described elsewhere.Birth outcomes of interest were neonatal death, infant death, preterm birth and birth weight. Neonatal death was defined as any death within the first 28 days after birth, infant death was defined as any death within the first year after birth  and preterm birth was defined as a birth before 37 completed weeks of gestation.Mother’s country of birth was the main explanatory variable. It was coded from the National Statistics Category Classification title of the country of birth reported in the birth registration data. Mother’s country of birth was further subclassified within categories of infant’s ethnicity as described below. Infant’s ethnicity, reported in the NN4B systemWe considered the potential for confounding by maternal and infant characteristics, year of birth and geopolitical circumstances of different countries and regions. The variables representing these factors in the data set were: maternal age, sex of infant, birth registration type , infant’s year of birth, area-based deprivation measured using the index of multiple deprivation (IMD) — based on the mother’s address — and whether the mother’s country of birth was considered a fragile state based on the World Bank ‘Low-income countries under stress’ and ‘fragile situation’ classifications.We calculated the absolute risks of neonatal death, infant death and preterm birth, and the mean birth weights across characteristics and overall. Crude and adjusted risk ratios for neonatal death, infant death and preterm birth were estimated using binomial regression, and the mean differences in gestational age-adjusted birth weight were estimated using linear regression, with UK-born groups as the reference category. We adjusted for maternal age (including a quadratic term for age to account for departures from linearity), birth registration type, year of birth, sex of infant, IMD decile and for Black African infants only, fragile state status of mother’s country of birth. Gestational age was fitted as a continuous variable in models that adjusted for it. The robust estimator of variance was used to allow for non-independence of birth outcomes among infants born to the same mother. All regression models were fitted separately for each ethnic group although pooled estimates and tests for effect modification by ethnic group are also reported. The analysis was conducted using Stata V.15.There were data on 4 744 666 infants in the linked data set provided by the ONS. After excluding 109 734 infants with implausible or missing birth weights or gestational ages, a total of 4 634 932 infants were eligible for inclusion in this analysisThe majority of infants were white (72.3%), either of British (64.9%) or other White ethnicities (7.4%). The remaining 27.7% infants included 12.5% of known minority ethnic groups, 9.1% mixed race infants and 6.2% infants whose ethnicity was not stated.More than three quarters (75.7%) of infants had UK-born mothers, with the remaining quarter approximately equally split between mothers born in their countries or regions of ethnic origin and those born elsewhere. The average age of all mothers was 29 years. Most birth registrations (53.9%) were by married parents and nearly a third (30.2%) were by unmarried cohabiting parents. The distribution of infants across deciles of IMD was uneven, with more infants born in more deprived areas.The highest absolute risks of neonatal and infant deaths were observed among Pakistani and Black Caribbean infants, and the highest absolute risks of preterm births were in Black Caribbean and Bangladeshi infants . BangladAbsolute risks of adverse outcomes and mean birth weight were lower among female infants compared with males, infants born in later years compared with the earlier ones, infants born to married parents compared with unmarried or single parents and infants born in less deprived areas compared with those born in more deprived areas.The majority of the mothers of White British infants (96.1%), Black Caribbean infants (63.3%) and infants whose ethnicity was not stated (73.5%) were born in the UK . BetweenThe overall risk of neonatal death was 2.1 per 1000, but it varied from 1.9 per 1000 in infants of UK-born mothers to 2.6 per 1000 in infants of mothers born in their countries or regions of ethnic origin . The higIn the ethnic group-stratified analysis, there was no evidence of a difference in risk of neonatal death among infants of non-UK-born mothers compared with UK-born mothers, except among Pakistani infants for whom the adjusted risk of neonatal death was 16% lower among the infants of Pakistani mothers born in Pakistan  and Black Caribbean infants whose mothers were born elsewhere for whom the adjusted risk of neonatal death was 71% higher than infants whose mothers were born in the UK .The infant death risk was 3.2 per 1000 overall but 3.0 per 1000 among infants of mothers born in the UK or elsewhere, and 4.1 per 1000 among infants of mothers born in their countries or regions of ethnic origin . PakistaWhen stratified by ethnic group there was no evidence of a difference in risk of infant death when comparing infants whose mothers were born in their countries or regions of ethnic origin to those whose mothers were born in the UK across all ethnic groups, except among Pakistani infants: the adjusted risk of infant death was 16% lower among infants of Pakistani mothers born in Pakistan compared to infants of Pakistani mothers born in the UK .The overall preterm birth risk was 5.6%, varying from 4.8% among infants of mothers born elsewhere, to 5.6% among infants of UK-born mothers . The absIn the ethnic group-stratified analysis, White Other, Indian and Pakistani infants of mothers born in their countries or regions of ethnic origin had lower adjusted risks compared with those whose mothers were born in the UK . Additionally, White Other and Pakistani infants whose mothers were born elsewhere had lower adjusted risks of preterm birth  than those of UK-born mothers.higher in infants of mothers born in their countries or regions of ethnic origin (95% CI 23 g to 27 g) and 6 g higher in infants of mothers born elsewhere (95% CI 4 g to 8 g) compared with infants of UK-born mothers. The ethnic group-stratified analysis showed that infants of non-UK-born mothers in all ethnic groups had higher adjusted mean birth weights than those of UK-born mothers, except White British infants of non-UK-born mothers whose gestational age-adjusted mean birth weight was 3 g lower (95% CI −5 g to −0.3 g) than that of UK-born mothers.The overall mean birth weight was 3.36 kg, varying slightly from 3.28 kg among infants of mothers born in their countries or regions of ethnic origin, to 3.39 kg among infants of mothers born in both the UK and elsewhere . BangladIt has previously been suggested that the infants whose ethnicity was not reported were probably White infants.We explored the variations in birth outcomes among infants of different ethnicities born in England and Wales between 2006 and 2012. We sought to determine the extent to which these variations in birth outcomes could have been explained by where the mothers of these infants were born. To achieve this we compared the outcomes between infants born to UK-born mothers, the majority and infants born to non-UK-born mothers, who made up about a quarter of the births in the data. We further distinguished non-UK-born mothers between mothers born in their country or region of ethnic origin and elsewhere. Such distinctions included, for example an Indian mother born in India, and an Indian mother born in Africa, both of whom then migrated to the UK and gave birth. Although we used ‘ethnic origin’ broadly for the purpose of classifying mother’s country or region of birth according to their ethnicity, we recognise that no single approach to this classification can fully encapsulate the nuances of ethnicity, nor of individuals’ perception of their own origins and heritage.Our pooled adjusted analysis showed evidence that the infants of non-UK-born mothers who were born in their countries or regions of ethnic origin had lower risks of neonatal death, infant death and preterm birth and higher mean birth weights than the infants of mothers born in the UK. There was evidence of a lower risk of preterm birth, but not neonatal or infant death, in the infants of mothers who were born neither in the UK nor in their country or region of ethnic origin , and evidence of a higher mean birth weight than in infants of UK-born mothers. The adjusted ethnic group-stratified analysis showed lower risks of adverse birth outcomes among the infants of non-UK-born mothers to be particularly evident among Indian (for preterm birth) and Pakistani infants , a higher risk of neonatal death among Black Caribbean infants whose mothers were born elsewhere and a positive effect on birth weight in all ethnic groups except White British infants.These findings are consistent with those of previous studies in England and Wales which found higher risks of adverse birth outcomes among minority ethnic infants relative to White British infants,The finding of evidence of higher risk of neonatal death among Black Caribbean infants of mothers born elsewhere stood in contrast with the pattern of findings of better outcomes among infants of non-UK-born mothers. A closer inspection found that the mothers of 64.9% of the non-UK-born Black Caribbean infants were born in Africa. This would imply that 11.8% of the non-UK-born Black Caribbean population between 2006 and 2012 was born in Africa, when in fact 2011 census data showed that only 1.3% of this group were born in Africa.Mother’s country of birth may influence birth outcomes of infants through conditions and risk factors which may be common to migrant groups, for example, health behaviour, cultural, environmental, economical, social and lifestyle factors,This study was based on a large number of births, which allowed us to conduct an ethnic group-stratified analysis with reasonable power to identify important differences. Our data set was drawn from national birth and death registration data and linked to routine hospital data to obtain covariates and additional outcomes not available in registration data, with few exclusions. It is therefore representative of England and Wales and may also apply to other high-income countries in Europe and possibly other countries with comparable health systems, socio-demographic characteristics and ethnic minority population distribution. However, our data did not include information about important risk factors for infant mortality, such as maternal smoking,This study shows evidence of better outcomes among minority ethnic group infants whose mothers were born outside the UK compared with those born in the UK, and highlights the heterogeneity across ethnic groups in the effect of mother’s country of birth on infants’ birth outcomes. If current migration and demographic trends continue into the future, increasing proportions of births in England and Wales will occur among non-UK-born women, the majority of whom belong to minority ethnic groups, and increasing proportions of the population will be of non-White ethnicity. With ethnic minority groups experiencing poorer birth  outcomes than majority groups, future reductions in the overall rates of adverse outcomes will require interventions that prioritise identification and targeting of at-risk groups. These findings will therefore inform risk-stratification strategies in public health interventions and policies aiming to improve birth outcomes.There is a disproportionate risk of adverse outcomes among infants with a minority ethnic heritage in the UK. Risk factors for adverse birth outcomes are generally well understood and specific causes of poorer outcomes in some ethnic groups are also well established. However, there is a dearth of studies exploring the joint effects of mother’s country of birth and ethnic group on birth outcomes across the range of ethnic groups in the UK.We highlight evidence of better birth outcomes among UK-born infants of non-UK-born mothers: infants of mothers born in their countries or regions of ethnic origin have better birth outcomes than those of UK-born mothers, but there is no evidence of a difference between birth outcomes for infants of mothers born elsewhere and those of UK-born mothers in most ethnic groups. We also show evidence of considerable heterogeneity across ethnic groups in the effect of mother’s country of birth on birth outcomes, with higher gestational age-adjusted mean birth weights in infants of non-UK-born mothers compared to UK-born mothers across all ethnic groups except White British infants, but lower risks of preterm births only among White, Indian and Pakistani infants, and lower risks infant or preterm birth among Pakistani infants. These findings could inform future healthcare interventions to identify and prioritise at-risk groups.


Female genital mutilation (FGM) has been a longstanding tradition in Egypt and until recently the practice was quasi-universal. Nevertheless, there are indications that the practice has been losing support and that fewer girls are getting cut. This study analyzes the prevalence of FGM in different birth cohorts, to test whether the prevalence declined over time. The study also examines whether such a decline is occurring in all segments of society or whether it is limited mostly to certain more modernized segments of society.This study pooled data from the 2005, 2008 and 2014 waves of the Egypt Demographic and Health Surveys (EDHS). The women participating in the EDHS provided data on 62,507 girls born to them between 1987 and 2014, including whether they were cut and at what age. Kaplan-Meier and Weibull proportional hazard survival analyses were used to examine trends in the prevalence and hazards of FGM across birth cohorts. Controls for region, religion and socioeconomic status of the parents were included in the Weibull regression.The results show a steady decline in FGM across the birth cohorts studied. The base hazard for the 2010 birth cohort is only 30% that of the 1987 one. Further analyses show that the decline in FGM occurred in all segments of Egyptian society in a fairly similar manner although differences by region, religion and socioeconomic status persisted.This study confirms that FGM is declining in Egypt. The proportion of girls getting cut has declined rapidly over the past few decades. This decline is not limited to the more modernized segments of society, but has spread to the more traditional segments as well. The latter increases prospects for the eventual eradication of the practice. Egypt Demographic and Health Surveys (EDHS) to examine the extent to which the prevalence of FGM has changed across the birth cohorts from 1987 to 2014. Using data on whether the daughters of the women participating in the EDHS were cut, and at which age they were cut, will provide information on more recent evolutions regarding FGM in Egypt.Egypt has a long history of female genital mutilation (FGM) and the practice is firmly embedded in Egyptian culture and tradition. Historically FGM had the support of almost the entire population and almost all girls were cut. The United Nations Children’s Fund (UNICEF)  reportedFGM is deeply embedded in a society’s culture and tradition and intimately related to women’s social status and honor –17. The However, although these customs reinforce the subordinate status of women they also provide status to women. Being cut symbolizes that a woman is of good standing, which tends to increase her marriageability, social status and prestige , 21–28. International organizations promote increasing female empowerment, improving women’s position in society, and reducing gender inequality as a strategy to eradicate FGM . TypicalGender Inequality Index declined slightly between 1995 and 2015, from 0.665 to 0.565, but it remains high and progress has been slower than in most other countries . However, even within birth cohorts there are variations in the risk of FGM. Modernized and marginalized groups function as innovators and will typically be the first to decide not to have their daughters cut. Girls born to better educated or higher socioeconomic status (SES) parents or from more marginalized  families will be less likely to be cut than those born in more traditional families [Hypothesis 2]. As anti-FGM attitudes diffuse throughout society, the risk of being cut is expected to decline in other segments of Egyptian society, thereby reducing the difference between the innovative groups and the rest of society, except for some lagging segments of society [Hypothesis 3].This paper uses data from the 2005, 2008 and 2014 EDHS –59. The N = 57,763 . However, the unit of analysis of our study consists of the women’s surviving daughters aged 0–19, about whom data about FGM was collected. The pooled surveys contain data on Nd = 62,507 daughters  from Nm = 37,409 mothers . This implies that the mothers in the sample had an average of 1.67 daughters; this varies little across survey waves . Most women (54.5%) only have a single daughter, 29.8% have two daughters, 11.2% 3, and only 4.5% four or more daughters. The procedures and questionnaires of all DHS surveys have been reviewed and approved by the ICF International institutional review board (IRB) and comply with the U.S. Department of Health and Human Services regulations for the protection of human subjects (45 CFR 46); they have also been reviewed by an Egyptian IRB to assure compliance with Egyptian rules and laws [https://dhsprogram.com.The target population of the EDHS surveys consists of ever-married women aged 15 through 49. The pooled sample size of the three surveys used is The dependent variable for this study is duration of exposure to the risk of FGM. For girls who experienced FGM, this variable equals the age (in years) at which they experienced FGM. For non-cut girls, it equals their age at time of the survey. A dichotomous status variable indicates whether the girl was cut. The year of birth of the girls was used to define the cohorts, and ranges from 1987 to 2014. For some analyses cohorts were combined in five-year intervals: 1987–1990, 1991–1995, 1996–2000, 2001–2005, 2006–2010, and 2011–2014. All confounding variables are measured at the level of the mother. These include the region where the mother lives, her religion, education level and occupation, as well as the education level and occupation of her spouse. In nearly all cases, the latter will be the girl’s biological father, but if the girl’s mother remarried, it may be her social father. For simplicity, we will refer to this person as the girl’s father.To track the evolution of FGM by age for each birth cohort Kaplan-Meier estimators are used with the larger five-year birth cohorts as stratifying variable. For the most recent birth cohorts the available data is limited to the younger ages, as the entire survival function cannot be estimated. Multilevel Weibull proportional hazard regression techniques are used to analyze the effects of the confounding variables on the risk of being cut. The regression coefficients represent the changes in the underlying log-hazard, i.e., they are partial log-hazard ratios. As before, for the younger birth cohorts only part of the survival function can be estimated. All analyses were conducted using the sampling weights provided by the EDHS surveys.Table The girls’ mothers also have become better educated over time. In the 2005 EDHS about 40% of the girls have an uneducated mother. In the 2014 EDHS this proportion has declined to 27%, while the proportion of girls with mothers with complete secondary or higher education has increased from 35 to 50%. Although better education for women is often seen as evidence of the improvement of women’s social position, in this case this improvement is clearly only partial as the labor market participation of the mothers remains poor and may even be worsening. In 2005 77% of the girls have non-working mothers, compared to 83% in 2014. More than 95% of the girls also have Muslim mothers. Pooled data from the three EDHS waves show that only 22% of girls have an uneducated father, while 60% have a father who has secondary or higher education. Only 3% of girls have a father who is not working and the two most prevalent occupational categories are skilled manual laborers (28%) and professional, technical or managerial occupations (23%).Figure Figure p < 0.001). The hazard ratio (HR) equals the antilog of the regression coefficients in Table HR(X) equals the HR associated with factor X and bX the Weibull regression coefficient for factor X. In Rural Lower Egypt the estimated HR is 3.25 (p < 0.001). The educational level of the mother is another important factor affecting girls’ risk of being cut. Girls born to better educated mothers are significantly less likely to be cut than those born to mothers without any formal education. The HR for girls born to uneducated mothers compared to girls born to mothers with a complete secondary education is only 0.36, while for those born to mothers with higher education it is only 0.10 (both p < 0.001). Noteworthy is that for girls born to mothers with an incomplete secondary education the hazard of being cut is slightly but significantly higher than for those born to mothers with no formal education . The labor market status of the mother also affects the risk of being cut. On average girls born to working mothers have a lower hazard of being cut than those born to non-working mothers. The only exception consists of girls born to mothers working in professional, technical or managerial jobs, whose hazard of being cut does not significantly differ from those born to non-working mothers. We also expected FGM to decline in the least traditional segments of society. The results show that girls born to Christian mothers are at a significantly lower risk of being cut than those born to Muslim mothers .Table The father’s background also affects whether a girl gets cut, independently from the mother characteristics, but these effects are considerably weaker than those of the mother’s characteristics. Girls whose father has either only primary education or has higher education have a lower hazard of being cut compared to when the father has no formal education, while if the father had secondary education the hazard of being cut does not differ significantly from those with no formal education. Regarding the occupation of the father, only girls whose father was working in agriculture had significant lower hazards of being cut, compared to those whose father was not working.Subsequent analyses examined whether the decline of the hazard for FGM is equal for all segments of society or whether certain groups lead the decline while others lag. The results are shown in Fig. For three of the confounding variables, mother’s religion, mother’s and father’s education, no significant interactions with the girl’s year of birth were observed, implying that the trend in the risk of FGM did not differ across the various categories of these variables and that they all followed the trend previously observed in Fig. For the other three confounding variables , significant interactions with the girl’s year of birth were observed, even after controlling for the other confounding variables, implying that the different categories not only differ in the level of risk but also experience different trends in the evolution of this risk. Although the pairwise comparisons of the trends for the different regions did not show any significant differences, Fig. The literature on FGM in Egypt suggests that the prevalence of FGM is declining and that an increasing proportion of girls no longer get cut , 60, 61.The results of this paper suggest that FGM in Egypt will keep declining in the near future. In just a few decades the number of girls who were subjected to FGM has already declined substantially, and once a tipping point has been reached a further decline may occur quite fast. However, success in eradicating FGM in Egypt is by no means a certainty. The final eradication of the practice will require a continued effort, not only from the public authorities enforcing anti-FGM legislation, but also from anti-FGM advocates in civil society. Further improvements in women’s social position may also contribute to the decline of FGM. Unfortunately, although more women are getting secondary or higher educated, their labor force participation is still lagging. In addition to the need to improve women’s social positions, the eradication of FGM also requires a cultural shift that severs the links between a woman’s honor, her family’s status and FGM. Egypt has made serious strides in this direction, but by no means has it already won the war.Additional file 1: Table S1. Weibull proportional hazard survival analysis results for age at FGM, including interaction terms between risk factors and birth cohort. Figure S1. Kaplan-Meier estimates for FGM by age, birth cohort, and region of residence. Figure S2. Kaplan-Meier estimates for FGM by age, birth cohort, and mother’s education. Figure S3. Kaplan-Meier estimates for FGM by age, birth cohort, and mother’s religion. Figure S4. Kaplan-Meier estimates for FGM by age, birth cohort, and mother’s occupation. Figure S5. Kaplan-Meier estimates for FGM by age, birth cohort, and father’s education. Figure S6. Kaplan-Meier estimates for FGM by age, birth cohort, and father’s occupation.


Are all-cause and noncommunicable disease (NCD) mortality risks higher amomg adults born preterm and early term compared with those born at full term?This cohort study of more than 6.1 million individuals from 4 Nordic birth cohorts provided found all degrees of preterm birth and early-term birth are associated with increased risk of premature adult death. Excess mortality associated with shorter gestation was pronounced for death from NCDs, such as cardiovascular diseases, chronic lung disease, and diabetes.These findings suggest that increased early adult mortality associated with shorter gestation includes excess risk in the large group of adults born late preterm and early term and is pronounced for NCD deaths. This cohort study examines the risk of all-cause and noncommunicable disease deaths among young adults born preterm and early term in 4 Nordic countries. Adverse long-term outcomes in individuals born before full gestation are not confined to individuals born at extreme gestational ages. Little is known regarding mortality patterns among individuals born in the weeks close to ideal gestation, and the exact causes are not well understood; both of these are crucial for public health, with the potential for modification of risk.To examine the risk of all-cause and noncommunicable diseases (NCD) deaths among young adults born preterm and early term.This multinational population-based cohort study used nationwide birth cohorts from Norway, Sweden, Denmark, and Finland for individuals born between 1967 and 2002. Individuals identified at birth who had not died or emigrated were followed up for mortality from age 15 years to 2017. Analyses were performed from June 2019 to May 2020.Categories of gestational age .All-cause mortality and cause-specific mortality from NCD, defined as cancer, diabetes, chronic lung disease, and cardiovascular disease (CVD).P for interaction = .03). Preterm birth was associated with 2-fold increased risks of death from CVD , diabetes , and chronic lung disease . The main associations were replicated across countries and could not be explained by familial or individual confounding factors.A total of 6 263 286 individuals were followed up for mortality from age 15 years. Overall, 339 403 (5.4%) were born preterm, and 3 049 100 (48.7%) were women. Compared with full-term birth, the adjusted hazard ratios (aHRs) for all-cause mortality were 1.44  for moderate preterm birth and earlier; 1.23  for late preterm birth; and 1.12  for early-term birth. The association between gestational age and all-cause mortality were stronger in women than in men (The findings of this study strengthen the evidence of increased risk of death from NCDs in young adults born preterm. Importantly, the increased death risk was found across gestational ages up to the ideal term date and includes the much larger group with early-term birth. Excess mortality associated with shorter gestational age was most pronounced for CVDs, chronic lung disease, and diabetes. So far, studies investigating long-term health and disease after preterm birth have mostly been limited to early adulthood among those with the highest risk, namely those born extremely (<28 weeks) and very (28-32 weeks) preterm.5 However, recent reports suggest that adverse long-term outcomes are not confined to extreme gestational ages, considering that children born just a few weeks before term or even early term had higher risk.6Globally, 15 million pregnancies each year, 1 in 10, result in preterm birth, ie, birth before 37 weeks of gestation.8 While many studies that have followed up individuals born in the mid to late 20th century demonstrate robust associations between lower birth weight and higher cardiometabolic disease risk and mortality,10 only some could assess preterm birth as a factor associated with these outcomes. Furthermore, results are mixed, and their generalizability to individuals entering adulthood today may be limited.12 A growing body of evidence from clinical follow-up studies associates shorter gestation with higher levels of risk factors for noncommunicable diseases (NCDs).15 For cancer, studies on associations with gestational age are few,16 while larger birth size has been associated with higher adult cancer risk.17The notion that early life is a vulnerable period, when even subtle disruptions in the development of organ systems may lead to adverse health outcomes in adulthood, is far from new.20 suggest that lower gestational age is associated with increased mortality in young adulthood. Norwegian data indicate similar patterns.21 The authors of the previous reports on Swedish data recently reviewed the evidence for preterm birth and adult mortality22 and found that studies from Sweden, Norway, and Australia had reported an association between preterm birth and adult mortality but that heterogeneity in analyses and reporting did not allow aggregation or any in-depth comparison of results.In most clinical follow-up studies of individuals born preterm, the participants are still too young for meaningful estimation of NCD risk and premature adult death. However, the nationwide medical birth registries established in the Nordic countries in the 1960s and 1970s enable follow-up into adulthood for large groups of individuals born preterm. Reports on gestational age and mortality in the Swedish population registriesHere, in what is, to our knowledge, the largest study to date and with a maximal follow-up until age 50 years, we expand previous findings in aggregated analyses with follow up among more than 6 million young adults from 4 Nordic countries to assess remaining questions and to justify power for analyses of specific NCD-related causes of death. Our aim was to investigate the association between gestational age categories across the entire gestational age spectrum and all-cause and NCD mortality in young adults in Norway, Sweden, Denmark, and Finland.23 in each country were linked with data from the respective national causes of death registries27 using unique identifiers. The study population included all children born alive who were recorded in the Norwegian MBR from 1967 to 2002, the Swedish MBR from 1974 to 2002, the Danish MBR from 1978 to 2001, and the Finnish MBR from 1987 to 1990. For each country, individuals were excluded if their gestational age was outside a plausible range ; if their birth weight was greater than 6000 g or less than 350 g; if their birth weight for gestational age was more than 6 SDs from expected28; if maternal identity could not be ascertained; or if they died or emigrated before age 15 years. Furthermore, individuals were excluded from the study populations if they had missing information on gestational age, birth weight, sex, birth year, and maternal age. The study followed the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) reporting guideline.This study was based on individual-level data from nationwide registries in Norway, Sweden, Denmark, and Finland. Data from the medical birth registries (MBRs)29 According to Danish legislation, no ethical permission is required for register-based research; however, this study was approved by the local data protection authorities. This study was approved by the Swedish ethical review board of Stockholm and the research ethics board of the Finnish Institute for Health and Welfare.The Norwegian ethical committee on medical research approved this study; data were delivered and approved by the Norwegian MBR and cause of death registry, both at Norwegian Institute of Public Health.30 In Denmark, gestational age was estimated based on last menstrual period, ultrasonography, or clinical examinations.31 In Finland, gestational age was estimated based on last menstrual period and, from the late 1980s onward, confirmed by ultrasonography examination. Gestational age in completed weeks was categorized as moderately preterm or earlier (23-33 weeks), late preterm (34-36 weeks), early term (37-38 weeks), full term (39-41 weeks), and post term (42-44 weeks). However, for underlying causes of death that are rare in this relatively young population , all individuals born before 37 weeks were analyzed as 1 preterm group.Exposure information on gestational age was obtained from the 4 MBRs. In Norway, gestational age was estimated from last menstrual period, ultrasonograph examination, or, in cases of assisted reproduction, from date of embryo transfer. In Sweden, early second-trimester ultrasonography examination has been routinely offered since 1990, and more than 95% of women attended; otherwise, the date of the last menstrual period was used.International Classification of Disease (ICD), Eighth, Ninth, and Tenth editions, and codes of primary underlying causes of death were grouped according to the European Shortlist for Causes of Death (EU short codes) version 1998.32 All-cause mortality and the following primary underlying causes of death were investigated: cardiovascular disease , divided into noncerebrovascular CVD  and stroke ; cancer , diabetes ; chronic lung disease , and NCD .Causes and dates of death were obtained from the causes of death registry in each country. Underlying causes of death were registered according to the Information on birth year, sex, birth weight, congenital malformations, maternal age, and parity was obtained from the national registries. Congenital malformations were defined as having 1 or more congenital malformations registered. Parity was defined as number of previous live births (0 and ≥1).33 for each comparison.34 Analyses were performed from June 2019 to May 2020.Survival time was calculated as time from age 15 years to death, emigration, or end of follow-up . The Norwegian and Swedish data were pooled and analyzed together; the Danish and Finnish data were analyzed separately and combined in a meta-analysis using the inverse-variance methodFor Norwegian and Swedish data, individual data could be pooled and analyzed in Norway. For Denmark and Finland, individual data could not be shared, and analyses had to be performed locally due to restrictions in data sharing regulations. Thus, in Norwegian and Swedish data, more sophisticated analyses, including sibling analyses and cumulative hazards with interaction of sex, could be performed.35 In cause-specific modeling, competing risks  were treated as censored.36 The proportional hazards assumption was assessed by plotting the log-cumulative hazard against the log of survival time. There was no imputation of missing data.Relative risk of mortality in categories of gestational age compared with children born in weeks 39 to 41 was estimated by calculating hazard ratios (HRs) and associated 2-sided 95% Wald type CIs from Cox regression. The HRs were calculated adjusting for potential confounding by maternal age, parity, birth cohort, sex, and birth weight SD scores. Variables were categorized as presented in eTable 1 in the P < .05, and all tests were 2-tailed.Sensitivity analyses were performed to assess robustness of main findings. Thus, to explore potential effects from including older populations with longer follow-up in the Norwegian and Swedish data, study populations were restricted to individuals born from 1974 (earliest Swedish cohort) and from 1978 (earliest Danish cohort) in sensitivity analyses. To investigate a potential interaction between sex and gestational age, an interaction term was included in models based on the joint Norwegian and Swedish study populations. To examine familial confounding in the association between gestational age and mortality, sibling analyses were conducted in the Norwegian and Swedish data for the main outcomes  using a stratified Cox model, with mother’s identification number defining the strata. Individuals with congenital malformations were excluded in an additional sensitivity analysis. Maternal educational level was included as an additional confounder in sensitivity analyses conducted in the Swedish and Danish study population. All data handling and analyses were performed in SAS version 9.4 (SAS Institute). Statistical significance was set at Among 6 263 941 individuals identified at birth, a total of 6 263 286 individuals  could be followed up after age 15 years. Overall, 339 403 (5.4%) were born preterm, and 3 049 100 (48.7%) were women.Detailed characteristics of the study population are presented in eTable 1 in the During follow-up, 46 936 (0.7%) individuals died. Median (range) age at death was 28.8 (15.0-50.9) years. A total of 10 430 (22.2%) deaths were from NCDs and 2880 (6.2%) deaths were from CVDs. The cause of death was not registered for 608 of 21 706 (2.8%) in Norway, 36 of 18 138 (0.2%) in Sweden, 204 of 5717 (3.6%) in Denmark, and 5 of 1375 (0.4%) in Finland. The most common cause of death category not included in the NCDs studied was external causes of death (accidents and suicides). The median (interquartile range) survival time for those who survived until the end of follow-up was 32.8 (23.6-43.0) years in Norway, 28.7 (22.6-36.3) years in Sweden, 25.5 (20.3-31.5) years in Denmark, and 26.1 (25.1-27.0) years in Finland.P for interaction = .03). The aHRs for all-cause mortality in the moderate preterm group or earlier compared with full term was 1.63  in women and 1.25  in men  for all-cause death after age 15 years in individuals born before week 34 compared with individuals born full term was 1.44 ; the corresponding aHR for those born moderately preterm and earlier was 1.23 , and for those born early term, it was 1.12  . Thus, aResults for NCD-related deaths were similar to those for all-cause mortality. A dose-response association across the preterm and early term groups was observed .Cancer mortality was modestly increased for individuals born moderately preterm and earlier as well as for the postterm group compared with the full-term group. However, precision was low owing to few cancer deaths, especially in the group born moderately preterm and earlier   similar results were observed. In the current study, we pooled updated Norwegian and Swedish data through 2017 and included Danish and Finnish birth cohorts in a meta-analysis approach to present the largest study to date to our knowledge, with a maximal follow-up until 50 years of age, to assess remaining questions and to justify power for analyses of specific NCD-related causes of death. The results are mainly homogenous across the countries displayed. Our data support the finding that the risk of premature death in individuals born preterm compared with those born full term is higher in women than in men. Because women with cardiometabolic risk patterns and less favorable socioeconomic conditions are at a higher risk of both pregnancy complications and preterm birth,37 it could seem plausible that the association of shorter gestation with adult mortality could be confounded by maternal and socioeconomic factors. Similarly, offspring with congenital malformations are more frequently born before term and have long-term morbidities that may influence mortality.39 Nevertheless, both the current and previous studies that have been able to take these factors into account21 indicate that the increased adult mortality in individuals born after shorter gestation cannot be attributed to any of these factors. A tendency of marginally strengthened estimates associated with preterm birth if birth years are restricted to births after the late 1970s (rather than including the late 1960s and the early 1970s) was evident for CVD. Weaker estimates for adult mortality in the groups with the most preterm birth in the earliest birth cohorts could be explained by lower follow-up rates due to higher child mortality in this group. Also, a higher proportion of gestational age estimations based on last menstrual period (rather than based on ultrasonography) in the earlier cohorts may decrease precision and thus attenuate associations for the earliest years.Swedish studies from 201140 By pooling analyses across the Nordic countries, we are in a unique position to address limitations of previous studies on surviving individuals who were born preterm, are still relatively young, and for whom death from NCDs is still a relatively rare outcome. The longitudinal nature of the Nordic registries and the opportunity to link data between the national registries enabled us to follow up individuals from birth into adulthood (to a maximum of 50 years) in 4 Nordic countries. The 4 large populations allowed analyses to include specific groups of NCD mortality. The use of nationwide population registries also reduced selection and ascertainment biases. The data from Norway and Sweden were combined on an individual level, and this enabled the investigation of familial confounding, the investigation of interactions between sex and degree of prematurity, and the investigation of rare causes of death. Findings from the Swedish and Danish population, where information on maternal education was available, did not suggest that socioeconomic status could explain the association of gestational age with adult mortality.In the current analyses, we have overcome regulatory barriers that often preclude aggregated analyses of sensitive registry data across countries.41 Consequently, this could have produced less contrast and more conservative estimates, particularly in the earlier years of the study period, when ultrasonography was not used routinely. Lower neonatal and childhood survival in the earliest birth cohort may have introduced survivor bias, with only the healthiest reaching adulthood. Neonatal survival and treatment has changed substantially from the beginning of the study period42; therefore, it is unclear to what extent our findings are generalizable to children born today and to children in other settings with fewer resources. It remains to be observed whether these patterns of higher mortality in individuals born before term persist later in adulthood and whether they apply to more recent birth cohorts with increased neonatal survival.There are important limitations to the study as well. The cohort is still relatively young, and despite a large sample size, the number of outcomes limited assessment of more narrowly defined preterm categories for the specific outcomes. Furthermore, different methods of estimating gestational age may have influenced the accuracy of the estimates of gestational age. Compared with estimates of gestational age based on last menstrual period, it has been found that estimates based on ultrasonography shifted the distribution of gestational age lower.The current findings add evidence of higher risk of young adult death from all causes and from chronic diseases in individuals born before full gestation. Importantly, the altered risk included the late preterm and early-term groups, who were born close to ideal birth timing. This may have implications not only for tailoring chronic disease risk assessments in individuals born before full term but also for the interpretation of the rich literature regarding the association between birth weight and mortality, in which gestational age has not been taken into account.


To examine the type of maternal dietary patterns during pregnancy and the distribution characteristics of children’s birth weight and the association between dietary patterns and neonatal birth weight in China.Data were derived from a cross-sectional program named “The prevalence and risk factors of birth defects in Shaanxi Province” in July to November in 2013. A stratified multistage random sampling method was used to select women and their children. The mother’s diet during pregnancy was investigated using semi-quantitative food frequency questionnaire (FFQ) to collect the frequency and amount of food consumption, and the newborn birth weight as well as related social demographic information was collected at the same time. In our study, 0–1 year old children and their mothers with complete dietary survey data were selected as research objects. The main dietary patterns were identified according to factor analysis, and latent class analysis (LCA) was used to investigate the social demographic factors affecting dietary patterns. The logistic regression model was used to assess the association between birth weight and maternal dietary patterns during pregnancy by establishing three adjusting models and the data were stratified for further analysis by urban-rural and regions.OR = 1.61, 95%CI:1.06–2.93) and middle region , and the traditional pattern had greater odds of lower birth weight in the middle region . The processing pattern was found a protective factor for the occurrence of low birth weight in rural areas  but was a risk factor for low birth weight in the southern region . The balance pattern was a protective factor for the occurrence of low birth weight in the northern region.A total of 15,980 participants were involved in this study. Four dietary patterns were identified: “vegetarian pattern”, “balance pattern”, “traditional pattern” and “processing pattern”. Compared with moderate tertile, women in the highest tertile of adherence to vegetarian pattern increased the risk of low birth weight in offspring in rural areas (The vegetarian and traditional pattern may be positively related to a higher risk of low birth weight while the balanced pattern may keep birth weight of offspring within the appropriate range. Health education of balanced diet and individual nutrition guidance during pregnancy should be strengthened, to make the dietary structure during pregnancy are more reasonable, reduce the occurrence of adverse birth weight of newborns. Pregnancy is a complex physiological process, the health of pregnant women and offspring is affected by nutrition during pregnancy. Reasonable and correct dietary structure is the guarantee for improving the quality of the population . The dieHowever, there are few research data in this aspect in China at present, and the dietary pattern has a strong local specialties and is influenced by many aspects. Chinese pregnant women have complex and diverse eating behaviors, and follow a set of dietary customs which are not extensively explored in the literature. For this reason, a large-scale population survey was conducted in Shaanxi, China. From a macro perspective, the main dietary patterns of women of childbearing age during pregnancy were analyzed, and the possible confounding factors were adjusted to further explore the relationship between dietary patterns and birth weight, SGA and LGA. Information about how maternal food quality is correlated to unfavorable birth weight can provide a basis for nutritional interventions to improve both maternal and infant health in western China.The data were derived from a cross-sectional program named “The prevalence and risk factors of birth defects in Shaanxi Province” ,which was the Sub-topics of National Natural Science Foundation of China project “The primary preventive effect of micronutrient intervention on the occurrence of congenital heart disease during perinatal period and its mechanism”, which were collected from July to November in 2013. A stratified multistage random sampling method was used to carried on sampling, random sampling of districts and counties was carried out after stratification by urban and rural areas. According to the urban-rural ratio and taking into account the population intensity and fertility level, 10 urban districts and 20 counties were randomly selected. The difference between urban and rural areas was determined according to the overall urban and rural population in Shaanxi, making the sample representative. The sampling methods were as follows: 1) 6 townships were randomly selected from the sample counties, and 6 villages were randomly selected from each township. Each village randomly surveyed 30 women aged 15–49 who were pregnant after 2010 and had a clear pregnancy outcome. 2) In each of the urban districts, 3 sub-district offices were randomly selected, and each district randomly selected 6 communities, each community randomly surveyed 60 eligible women of childbearing age. The subjects included women of childbearing age who had been pregnant between January 2010 to November 2013, met the inclusion criteria, had known outcome and informed consent. The inclusion criteria were that women were pregnant from 2010 to 2013, had no major diseases, were willing to participate in this study, and completed the questionnaire independently under the guidance of the investigator. The exclusion criteria were those with twin or multiple pregnancies, with severe memory impairment of dietary intake during pregnancy, and those with diabetes mellitus, severe heart, liver and kidney diseases and psychosis and impairment of understanding ability during gestation. The investigation was approved by the Medical Bioscience Research Ethics Committee of the Ministry of Medicine of Xi’an Jiaotong University (No. 2012008), and all respondents signed informed consent. In order to reduce the error caused by recall bias and increase the accuracy of the results, this study selected children aged between 0 and 1 year old and their mothers with a complete dietary survey data as subjects.Data collection was conducted using the questionnaire “Status of Birth Defects and Risk Factors in Shaanxi Province”. The content of questionnaire included: basic conditions of the mother, birth weight of the newborn, sociodemographic characteristics , etc., food frequency questionnaire (frequency and amount of food consumed by the mother during most recent pregnancy).Food intake was conducted by trained investigators using the semi-quantitative food frequency questionnaire (FFQ) . Conside10) of the average gestational age; infants with birth weight above the 90th percentile (P90) of the same gestational age were defined as larger for gestational age (LGA); between P10 and P90 of the same gestational age was appropriate for gestational age (AGA) [Information about the birth wight of the newborn was collected mainly through birth certificates. When the newborn weight < 2500 g, it was defined as low birth weight infant; when the newborn weight ≥ 4000 g, it was defined as macrosomia; when the newborn birth weight 2500 ~ 4000 g, it was defined as normal weight infant . Referrige (AGA) .Before the implementation of the investigation, a rigorous scientific research design was carried out, and a training manual for investigators was prepared, all investigators received unified training. A community or village was selected for pre-investigation to ensure that each investigator correctly understood each item of the investigation form, and participated in the on-site investigation with a certificate of quality. The completed questionnaires were collated and verified by the investigators. The questionnaires that didn’t answer according to the regulations and had too many missing items were defined as invalid questionnaires, and the invalid questionnaires were excluded. After each county (district) completed the on-site investigation, 5% of the subjects were randomly selected for repeated investigation, and the results were reproducible to ensure the authenticity of the survey results.1 ~ Q3); categorical variables were described by composition ratio and rate. For missing values, we used the mean padding method for processing. These analyses were performed with the use of SAS.9.1, SPSS18.0, Mplus6.0, the drawing process was carried out by Microsoft Office 2003 Excel and Visio software. P values (two sided) < 0.05 was considered statistically significant.The questionnaire data were inputted by Epidata 3.1, and the data was double-input. Due to regional differences in intake of food and nutrient, the analysis was conducted stratified according to women’s location  and region. The continuous variables were statistically described by mean  and factor analysis . All 31 OR and 95%CI were calculated. Covariates included infant gender; mother’s age and gestational age, mother’s residence; education level; husband’s residence; family monthly expenditure; regional classification. Further stratified analysis according to the residence  and regional .All participants received factor scores for each dietary pattern, the scores of each dietary pattern were categorized into tertiles (Tertile1 (low score group), Tertile2 (middle score group) and Tertile3 (high score grouping)), the middle tertile (T2) score of each dietary pattern was considered to be the reference. The birth weight of the newborn was used as the dependent variable, and each classified individual dietary pattern was as an independent variable, the logistic regression model was used to investigate the association between birth weight and maternal dietary patterns during pregnancy, and the crude and multivariate-adjusted Firstly, Lo-Mendell-Rubin likelihood ratio test was used to determine the optimal number of LCA selections and compared with dietary patterns after extracting four dietary patterns by factor analysis. The most widely used signal evaluation indicators in the LCA was AIC criterion (akaike information criterion) and the BIC criterion (bayesian information criterion), both indicated that the smaller the value, the better the fit. In the LCA analysis, every women could predict a certain type of probability, and subjects could be grouped into the type of dietary pattern with the highest probability. Then the study mainly used the LCA method to calculate the intake of various foods in each dietary pattern, compared the sociodemographic characteristics and nutrient intake between the various dietary patterns, multiple tests were compared by anova, mann-whitney test and Bonferroni correction.A total of 15,980 pregnant women who met the requirements were included in the study, aged 17 to 47 years (27.6 ± 4.5). There were 8501 (53.2%) male infants and 7479 (46.8%) female infants; they were aged between 0 to 12 months (6.6 ± 2.6), and the birth weight was 1050-6250 g (3279.9 ± 454.6 g). The basic characteristics of the research object were shown in Table In this study, the average weight of newborns in urban areas was 3339.2 ± 461.6 g, 3230.7 ± 451.0 g in rural areas, 3263.7 ± 453.3 g in southern Shaanxi, 3290.3 ± 464.7 g in northern Shaanxi, and 3252.6 ± 450.9 g in middle region. When there was no stratification, 783 newborns with birth weight of < 2500 g, accounting for 4.9%, and 703 newborns with birth weight > 4000 g, accounting for 4.4%; the incidence of SGA and LGA was 14.7 and 7.6%, respectively  and BIC (1311.01) were the smallest, the optimal model was obtained. This result was consistent with the classification of dietary patterns obtained by factor analysis. Figure 307.28 anP = 0.018), the higher the monthly household expenditure, the more inclined the balance pattern and processing pattern, while the lower ones were more likely to choose the vegetarian pattern. Urban women were more inclined to choose balanced pattern, rural women were more inclined to choose vegetarian pattern (P = 0.020). The higher the education level of the mother, the more inclined to choose the balanced pattern, the lower the education level, the more inclined to vegetarian pattern and processing pattern(P = 0.035). Women with gestational age of less than 37 weeks tended to choose the traditional model, while women with gestational age of more than 37 weeks tended to choose the balanced model (P = 0.003). Women in southern region mostly relied on balanced pattern and processing pattern, women in northern region were mostly based on traditional pattern, while women in middle region were mostly vegetarian pattern (P < 0.001) ; balanced pattern (average protein intake of 78 ± 9 g) was more adequate than the other three groups of protein intake (P < 0.001); processing pattern (average carbohydrate intake was 389 ± 133 g) and traditional pattern (average carbohydrate intake was 377 ± 137 g) compared to the balanced pattern (average carbohydrate intake was 349 ± 132 g) and the vegetarian pattern (average carbohydrate intake was 369 ± 126 g), were more likely to intake more carbohydrate (P = 0.040); processing pattern (average cholesterol intake of 220 ± 105 g) was higher than the other three groups of cholesterol intake (P < 0.001); balanced pattern (average vitamin A intake was 637 ± 452μgRE) was more adequate than the other three groups of vitamin A intake (P = 0.026); balanced pattern  was higher than the other three groups of vitamin B12, folic acid, calcium intake (P < 0.001); the balanced pattern (average iron intake was 33 ± 12 mg) was more adequate than the other three groups of iron intake (P = 0.044). For other nutrients, there was no difference between the four dietary patterns  and balanced pattern  compared to traditional pattern  and vegetarian pattern , more easy to intake more energy , women in the highest tertile increased the risk of low birth weight in offspring , and the results remained unchanged after adjusting for covariates. In rural areas, vegetarian pattern was positively associated with a higher risk of low birth weight , the processing pattern was found a protective factor for the occurrence of low birth weight . The same results were not shown in urban areas.Unstratified analysis, women in the lowest tertile of adherence to vegetarian pattern had lower risk of low birth weight in offspring compared with moderate tertile . In northern Shaanxi region, the balanced pattern T1 group could increase the risk of low birth weight , the T3 group reduced the risk of low birth weight . In middle region, women in the highest tertile of adherence to vegetarian pattern had higher risk of low birth weight in offspring . The traditional pattern T1 group could reduce the risk of low birth weight in offspring , the T3 group increased the risk of low birth weight   showed that pregnant women who adhered to a prudent diet during pregnancy  had a lower risk of preterm labor than other pregnant women [This study found that in the unstratified and stratified rural areas and middle region, the incidence of low birth weight of newborns was positively correlated with the score of vegetarian-pattern diet, and the pregnant women with high scores in the traditional pattern also had an increased risk of low birth weight in middle region. Women who adopted these two dietary patterns were prone to cause insufficient intake of energy, protein, essential fatty acids, fat-soluble vitamins, and minerals rich in animal foods such as iron, calcium, and zinc, resulting in fetal growth disorders and affecting normal organ and tissue functions. However, a study in New Zealand showed that the higher the scores of traditional patterns  in early pregnancy, the lower the incidence of low birth weight , possiblnt women . Althougnt women –39.In summary, the dietary pattern of women of childbearing age may be closely related to the birth weight of the offspring. In order to reduce the occurrence of abnormal birth weight, women should adhere balance pattern during pregnancy, increasing the intake of high quality protein food, especially milk and seafood. It is suggested that dietary intervention during pregnancy should be improved, health education for dietary nutrition should be strengthened, and women should be able to make reasonable meals according to the actual local food supply characteristics. The advantage of this research is that the survey has undergone rigorous scientific design, the quality of the investigators is high, and the on-site work is organized and planned, the collected data is relatively accurate. The sample size of this study is large, and the possible confounding factors are controlled as much as possible during the analysis, increasing the reliability of the conclusion. However, there are still some limitations in this study. First of all, this study mainly adopts the cross-sectional research method, which cannot verify the causal relationship. Secondly, in this study, dietary information during pregnancy was recalled by subjects and collected through FFQ, which suggests that the recall bias could not be avoided. Thirdly, it is subjective to define the dietary pattern according to the concentration of food types contained in the dietary pattern, but the FFQ survey still has important reference value for the study of the dietary structure and food types of the population, especially the factor analysis plays a good role in the classification of dietary pattern –43. FinaThere are four kinds of dietary patterns in the pregnant women of childbearing age in Shaanxi Province, “vegetarian pattern” is predominant, and the dietary structure needs to be improved. The vegetarian and traditional pattern may be positively related to a higher risk of low birth weight while the balanced pattern may keep birth weight of offspring within the appropriate range. Health education of balanced diet and individual nutrition guidance during pregnancy should be strengthened, in order to make dietary structure during pregnancy are more reasonable, reduce the occurrence of adverse birth weight of newborns.