Background: There is wide variation in the macronutrient ingredient base of infant formula. How variation in macronutrient ingredients may impact infant growth remains largely unknown. Methods: The 2015–2016 National Health and Nutrition Examination Survey (NHANES) dataset was utilized, including infant anthropometrics and dietary intake. The protein, fat, and carbohydrate sources of formulas consumed were assembled and considered as potential predictors in multivariable models of infant Z-scores among infants < 6 months, 6–12 months and all infants combined (0–12 months). Results: The following relationships represent ingredient covariates within the final multivariable models of infant Z-scores. Consuming formula with palm oil was associated with higher weight-for-length Z-scores among infants < 6 months, but lower weight-for-age and weight-for-length Z-scores among infants 6–12 months. Consuming soy-protein formulas was associated with lower weight-for-length, head circumference-for-age and abdominal circumference-for-age Z-scores among infants < 6 months. Consuming sucrose-containing formula was associated with higher weight-for-length and abdominal circumference-for-age Z-score among infants 0–12 months. Conclusions: These data provide proof-of-concept that all formulas are not the same. Variation in macronutrient ingredients within the standard formula category is associated with differences in infant anthropometric outcomes. Long-term and mechanistic studies are warranted to pursue these findings; especially for palm oil, soy protein, and sucrose. In the United States, the composition of the 29 essential nutrients in infant formula is regulated by the Food and Drug Administration, and all sources of macronutrients used in infant formulas are approved for safety and maintenance of “normal” growth over the first year . VariatiHistorical research regarding protein source in formula has primarily focused on soy vs. dairy-based formula ,3,4, andInfant formulas in the US provide fat as a blend of vegetable oils. The focus of previous research regarding infant formula fat sources has been on the effect of palm oils on calcium and fat absorption, and bone mineral accrual ,20,21,22The caloric carbohydrate source in breast milk is lactose. Often alternative glucose or sucrose-based carbohydrates are utilized in infant formula as a replacement for lactose. Recently, a single study of 23 formula fed infants reported that consuming a glucose-based sugar (corn syrup solids) in formula, results in significant alterations in postprandial measures of metabolism, compared to lactose [Give the limited research investigating variation in macronutrient composition of infant formulas on the market and infant growth measures, our aims were to: document the variation in macronutrient intake of a nationally representative sample of formula fed infants in the United States and to investigate any relationships between macronutrient ingredients and infant anthropometric measures.Infant formula macronutrient ingredients and relative proportion of ingredients for all name brand formulas on the US market in 2015–2016 were assembled utilizing product websites, and contacting the company when information was unavailable either on the public or medical practitioner websites.Protein source was classified as % casein and % whey for dairy-based formulas and categorically yes/no for soy-protein based formulas. Protein size was categorized as either: amino acid, fully hydrolyzed, partially hydrolyzed, or intact protein. Caloric carbohydrate source was classified as a percentage of three categories based on the metabolic byproduct of the ingredient used: lactose, glucose, and sucrose. Corn syrup, corn syrup solids, brown rice syrup, starch, glucose syrup, and maltodextrin were all considered “glucose”. Formulas were also classified categorically as containing sucrose or not. Fat source was classified by percentage of fat that was provided by the following categories: soy oils, high oleic sunflower or safflower oils, and coconut or medium chain triglycerides (MCT) oils. Formulas were also classified categorically as containing palm or palm olein oils or not.Data from the National Health and Nutrition Examination Survey (NHANES) 2015–2016 were utilized. Dietary data and clinical data  were downloaded from the public website . The folInfant diet data were averaged when two days of dietary intake were available. Average daily caloric intake from formula, average number of formula feeds and average number of human milk feeds were calculated. Breastfeeding exclusivity was calculated as the proportion of total breast milk and formula feeds that were breast milk (i.e.: 100% = exclusively breastfed and 0% = exclusively formula fed). Breastfeeding exclusivity was also categorized as exclusively breastfed, vs. mixed-fed, vs. exclusively formula fed. Birth weight was categorized as follows: <2500 g = low birth weight (LBW) and >4000 g = large for gestational age (LGA). Calculated and relevant clinical NHANES variables from each dataset were merged by unique identifier. Only data from infants ≤ 12 months at the time of NHANES exam were kept in the dataset. Formula nutrient composition was merged into this trimmed dataset based on formula name.Non-normally distributed variables were either log or square-root transformed before analyses. Simple linear regression (univariate) analysis was used to determine if any of the macronutrient ingredient variables were correlated with infant z-scores. Breastfeeding exclusivity category, total daily kcal intake from formula, consuming solid food (or not) and birth weight category were also considered as potential predictors of Z-scores.p ≤ 0.2 were entered as independent variables into a multivariable model of Z-scores. Backward stepwise regression was then utilized to achieve the most parsimonious model that maximized the model’s adjusted R2 (which are reported) with the minimum number of covariates. The normality of all models’ residuals was confirmed.Variables that were independently associated with Z-scores at a significance level of This modeling was performed among infants < 6 months of age as a means of focusing on infants who’s primary source of calories was breast milk and/or formula, and among older infants (6–12 months) and on the population as a whole .Head circumference and abdominal circumference was not collected on every infant. Head circumference was only measured on 13 infants between 6–12 months of age. Thus, HCZ was only modeled for infants < 6 months of age.Results are reported as means ± standard deviations (SD) unless otherwise noted. For covariate variables that were transformed within models, parameter estimates reported herein were back-transformed in order to be clinically interpretable. Analyses were performed in SAS 9.4 and JMP Pro 14 Sample size numbers for the final dataset generation are presented in Cohort characteristics are presented in Among infants who were consuming sucrose (15.4% of infants), their carbohydrate intake consisted of 22.4 ± 7.8% sucrose. No infants were consuming premature infant formula.A total of 230 infants ≤ 12 months of age were consuming some formula. Parsimonious multivariable models of each Z-score outcome were generated for infants < 6 months, 6–12 months, and all infants combined (0–12 months) and are presented in A total of 124 infants < 6 months of age were consuming some formula.p < 0.2. In the final model of WAZ, only birth weight category remained in the model.In univariate models of WAZ, protein size , consuming palm oil, % fat as coconut oil/MCT, and % fat as soy oil, and birth weight category were independent predictors at p < 0.2.In univariate models of LAZ birth weight category was the only predictor at p < 0.2. In the final multivariable model of WLZ, consuming a partially hydrolyzed protein or consuming soy protein were associated with lower WLZ, while a fully hydrolyzed protein, consuming palm oil, and total daily kcal formula consumed were all associated with higher WLZ , consuming soy protein, consuming palm oil, % fat as coconut oil/MCT, % of fat as soy oil, total daily kcal formula consumed and breastfeeding exclusivity were independent predictors at gher WLZ .p < 0.2. In the final model of HCZ, consuming soy protein was associated with a lower HCZ.In univariate models of HCZ, consuming soy protein, total daily kcal formula consumed, and birth weight category were independent predictors at p < 0.2. In the final multivariate model of ACZ, consuming a soy protein formula and the % fat as coconut/MCT oil was associated with lower ACZ, whereas exclusive formula feeding (no human milk) was associated with higher ACZ.In univariate models of ACZ, consuming soy protein, consuming palm oil, % fat as coconut oil/MCT, % fat as soy oil, total daily kcal formula consumed and breastfeeding exclusivity were independent predictors at A total of 106 infants 6–12 months of age were consuming some formula.p < 0.2. In the final multivariable model of WAZ, birth weight category was associated with WAZ and consuming palm oil was associated with a lower WAZ.In univariate models of WAZ, consuming palm oil, % fat as soy oil, as well as birth weight category and whether the infant was consuming solid foods were all predictors at p < 0.2.In univariate models of LAZ, birth weight category was the only predictor at p < 0.2. In the final multivariable model of WLZ, birth weight category was associated with WLZ and consuming palm oil was associated with a lower WLZ.In univariate models of WLZ, consuming sucrose, consuming palm oil, birth weight category, and whether the infant was consuming solid foods were all predictors at p < 0.2. In the final multivariable model of ACZ, birth weight category was associated with ACZ and consuming sucrose was associated with a higher ACZ.In univariate models of ACZ, % of protein as casein, consuming sucrose, % fat as coconut oil birth weight category, and whether the infant was consuming solids were independent predictors at A total of 232 infants ≤ 12 months of age were consuming some formula. Parsimonious multivariable models of each Z-score outcome were generated and are presented in p < 0.2.Of all the variables considered in univariate models of WAZ and LAZ, birth weight category was the only predictor at p < 0.2. In the final multivariable model of WLZ, birth weight category was a significant predictor. Consuming a formula containing sucrose, and the proportion of fat consisting of soy oil were both associated with a higher WLZ.In univariate models of WLZ, consuming sucrose, % of fat as soy oil and birth weight category were independent predictors at p < 0.2. In the final multivariable model of ACZ, birth weight category was a significant predictor. Consuming a soy protein formula was associated with lower ACZ, whereas consuming a formula with sucrose and exclusive formula feeding (no human milk) were both associated with higher ACZ.In univariate models of ACZ, consuming soy protein, consuming sucrose, breastfeeding exclusivity, and birth weight category were independent predictors at These novel data suggest that standard variation in macronutrient ingredient sources in infant formula may be related to infant growth outcomes. Novel associations include the relationship between consuming sucrose and elevated WLZ and ACZ. However, the limitations of the dataset hinder the reach of conclusions. These data do provide proof-of-concept that variation in formula macronutrient source may impact infant outcomes. The relationships presented make a strong argument that mechanistic longitudinal investigations into each of the ingredients is warranted, and that standard infant formulas should not be considered “all the same”.Among infants < 6 months, consuming formulas with palm oil was associated with a higher WLZ. It is well documented that palm oil decreases intestinal calcium and fatty acid absorption in infants ,22,27,28A novel finding in this dataset was that the consumption of formula with sucrose as a carbohydrate correlated with higher ACZ among infants 6–12 months and higher WLZ and ACZ in infants 0–12 months. Sucrose has a higher glycemic index than lactose. Sucrose, a dimer of glucose and fructose, metabolizes to 50% fructose. Thus, in an infant exclusively consuming a formula with 27% carbohydrate as sucrose and 10.4 g carbohydrate/100 mL (a common formulation), 5.6% of the infant’s calories derive from fructose (via sucrose metabolism). Long-term consequences of infant consumption of sucrose is unknown. In adults, high fructose consumption has been associated with increased liver strain and can contribute to development of non-alcoholic fatty liver ,30 and mIn these data, consuming a soy protein formula was associated with a lower WLZ, HCZ, and ACZ in infants < 6 months, as well as a lower ACZ among all infants (0–12 months). Previous work has shown that infants fed soy formula have different body composition trajectories, exhibiting higher total fat free mass than breastfed and dairy-formula fed infants at 3 and 6 months but lower % fat mass than breastfed infants only up to 6 months . These aIt is interesting that these data do detect the previously reported observation of reduced WLZ among infants consuming partially hydrolyzed formula  among inOther relationships detected are thought provoking, including the positive relationship between amount of soy oil in formula and WLZ (in infants ≤ 12 months). This is clinically relevant as currently all US infant formulas in this dataset include soy oil in the fat blend at varying proportions.A significant limitation of this dataset is the cross-sectional design. This does not allow for control for infants’ previous Z-scores (i.e.: where they are on the growth curve). Additionally, gestational age is not available in this dataset. All models did control for birth weight category which is a loose proxy for size at birth but does not remove the limitation. While no infants in this dataset were consuming premature infant formula, we are unable to control for infants born prematurely, but advanced enough in growth to be transitioned to standard formula. The 0–12 month age group contained both the <6 and 6–12 month group, and thus are not independent models. Additionally, these data are from the United States and popular infant formula ingredient sources may vary internationally. The strength of some of the relationships is modest which contributes to the limitations. Strengths include representation from a national population, a large sample size, and the novelty of the incorporation of formula macronutrient formulation. This is also a novel presentation of the macronutrient sources of formula fed infants’ diets.In conclusion, these data provide evidence that the nutritional makeup of an infant’s formula needs to be included in considerations of growth outcomes in a clinical setting. Variation in ingredients within the standard formula category may indeed impact growth outcomes. Future research is needed to document the mechanistic effects of particular ingredients, and study long-term outcomes. Such studies are particularly needed for palm oil, sucrose, and soy protein.


Pest Management Science published by John Wiley & Sons Ltd on behalf of Society of Chemical Industry.Bio‐communication occurs when living organisms interact with each other, facilitated by the exchange of signals including visual, auditory, tactile and chemical. The most common form of bio‐communication between organisms is mediated by chemical signals, commonly referred to as ‘semiochemicals’, and it involves an emitter releasing the chemical signal that is detected by a receiver leading to a phenotypic response in the latter organism. The quality and quantity of the chemical signal released may be influenced by abiotic and biotic factors. Bio‐communication has been reported to occur in both above‐ and below‐ground interactions and it can be exploited for the management of pests, such as cyst nematodes, which are pervasive soil‐borne pests that cause significant crop production losses worldwide. Cyst nematode hatching and successful infection of hosts are biological processes that are largely influenced by semiochemicals including hatching stimulators, hatching inhibitors, attractants and repellents. These semiochemicals can be used to disrupt interactions between host plants and cyst nematodes. Advances in RNAi techniques such as host‐induced gene silencing to interfere with cyst nematode hatching and host location can also be exploited for development of synthetic resistant host cultivars. © 2020 The Authors.  Graphical representation of abiotic and biotic factors influencing the hatching and host seeking stages of cyst nematodes and approaches for use of semiochemical‐ and RNAi‐based management tools. These signals include visual, auditory, tactile and chemical, and can function in single, binary and ternary interactions both above‐ and below‐ground in a wide range of organisms; animals, insects, plants, microorganisms among others.Globodera and Heterodera, including potato cyst nematodes , soybean cyst nematodes (Heterodera glycines), sugar beet cyst nematode (H. schachtii) and cereal cyst nematodes (Heterodera avenae and H. filipjevi) among others.Cyst nematodes cause significant crop production losses worldwide. The most detrimental species of cyst nematodes belong to the genera Cyst nematodes attack essential crops, such as potatoes, soybeans, beans, rice, wheat, and barley, among others. Previously, it was thought that cyst nematodes existed only in temperate regions,H. glycines.Cyst nematodes management methods include crop rotation, fallow, trap crops, organic soil amendments, solarization, intercropping, chemical nematicides, biological control agents and genetic resistance. These methods have been used with varying degrees of success, while each is subject to various limitations. For instance, some synthetic nematicides have high mammalian toxicity and pose an adverse environmental hazard, hence their withdrawal from markets.To improve on the existing management strategies, it is critical to understand the abiotic and biotic factors, and the physiological mechanisms that influence bio‐communication in cyst nematodes. A recent review discussed the influence of root metabolites in the performance of soil nematodes.2G. rostochiensis is composed of mainly proteins, lipids, carbohydrates, glucosamine, polyphenols and inorganic matter.Cyst nematode hatching may be simple or complex depending on the nematode species. A cyst can contain hundreds of eggs, which are surrounded by the cyst wall. The cyst wall is a robust but permeable structure that acts as an interface between the nematode and its external environment. The cyst wall of G. rostochiensis as an example, a species that is almost entirely reliant on potato and tomato root diffusates to hatch. The first juvenile stage (J1) develops from an embryo, before molting into the J2 within the egg. Hatching of the J2 involves physico‐chemical alteration of the eggshell structure comprising three layers: outer lipoprotein layer, middle chitinous layer and an inner semi‐permeable lipid layer.G. rostochiensis usually occur in a state of partial hydration due to osmotic pressure generated by the disaccharide trehalose. Trehalose is found in the perivitelline fluid between the unhatched J2 and the eggshell and it protects the J2 against environmental shocks by inducing quiescence and inhibiting motility.H. schachtii, have a lower osmotic pressure due to low trehalose concentration enabling eggs to readily hatch in water without the need for host cues.The mechanism described in this section primarily uses G. rostochiensis, a calcium‐mediated process alters the permeability of the eggshell, increasing signalling of hatching factors. This process terminates diapause in the eggs and allows for an influx of water into the egg to hydrate the J2. This initiates a series of metabolic changes within the egg.In Physiological studies have revealed that the hydrated J2s emerge from the egg through a slit in the eggshell, which they create by thrusts from their stylets, allowing them to exit the cyst through the vulval fenestra or anus.e.g. sex pheromones) in the rhizosphere.Caenorhabditis elegans, have been shown to be multifunctional and each neuron bears more than one type of receptor.After hatching, cyst nematode J2s rely on lipid reserves as their sole source of energy until they locate a host plant, penetrate its root, and establish the syncytium (feeding site) in the vascular cylinder. The J2s use chemical and tactile cues as primary signals for host plant detection, orientation, guidance and communication  among other functions within the rhizosphere Fig. . Cyst nee.g. flowering, tuber formation) leading to a quantitative or compositional change in root diffusates. The stimulatory effect of root diffusate on hatching is dose‐dependent, often with optimal stimulation achieved at relatively low doses. At high doses, root diffusates can cause supra‐optimal inhibition, which is an intrinsic characteristic of hatching factor‐receptor interaction at high ligand concentration. Low doses of root diffusates resulting in high levels of hatch stimulation make a promising and practical target for use in exogenous application stimulation of ‘suicidal hatching’. ‘Suicidal hatching’ is the stimulation of cyst nematode hatching that result in eventual death of the nematode due to the absence of host plants. This can be achieved in various ways such as use of trap crops that can stimulate hatching but which do not allow nematode development and reproduction; exogenous application of host root diffusates; and use of synthetic hatching factors that do not allow hatched J2s to migrate from the cyst.Root diffusates from young plants tend to have a higher stimulatory effect on hatching than those from more aged/mature plants, attributed to either variations in the composition or concentration of the root diffusates. This demonstrates an adaptation of cyst nematodes to perceive the most suitable vegetative phase of host plants for optimum infection and survival.H. cajani, encysted eggs during a growing period was similar over the first four generations, but in the fifth and sixth generations, hatching was highly dependent on host root diffusates.The changing vegetative state of a plant during growth can also influence cyst nematode dependency on host root diffusates for hatching. For example, hatching of pigeon‐pea cyst nematode, Phaseolus vulgaris) stimulates hatching of H. glycines,Solanum lycopersicum) and potato (Solanum tuberosum) root diffusates stimulates hatching of potato cyst nematodes , in potato root diffusates was detected even though the structures of these hatching factors were not identified.G. rostochiensis to these hatching factorsSome of the host derived hatching factors include polar compounds: the pentanortriterpene glycinoeclepin A, isolated from the root extract of the kidney bean  was first reported to occur in the free‐living nematode Agrobacterium rhizogenes hairy root transformation and the gateway cloning system for rapid screening of candidate genes,Because cyst nematodes rely heavily on semiochemicals from host plants for cues influencing hatching and host‐seeking behavior, RNAi approaches to suppressing gene transcription in signal reception and transduction present powerful discovery tools in defining these interactions. RNAi has recently been utilized to define potential functions of novel cyst nematodes effectors,C. elegans and potato cyst nematodes, neprilysins are strongly expressed prior to hatching. There are over 20 neprilysin genes in C. elegans, while 11 different transcripts were up‐regulated in potato cyst nematodes in response to hatching factor. Homologs of both lap (lap‐1) and chitinase (cht‐2) have been identified in the potato cyst nematodes genome using C. elegans queries. Among the genes reported to be up‐regulated by hatching factor‐stimulus, many are components of calcium signaling and transport pathways, including gcy‐9 (guanylate cyclase/peptide receptor), mua‐3 , a gene encoding a DEL protein , and numerous transmembrane transporter genesUp‐regulated genes during hatching represent potential targets for RNAi to determine their functional roles in recognition of semiochemical signals and cellular processes involved in hatching. Candidates for RNAi screening for a reduced or suppressed hatching phenotype include the enzymes neprilysin , and chitinase, identified as cellular process components in previous studies. In C. elegans, much is known about the neurobiology of these systems in free‐living nematodes. As there is limited knowledge on which neurosensory systems in plant parasitic nematodes play a role in host recognition, identification of homologous systems using C. elegans as a guide is a suitable starting point. The chemosensory system in C. elegans is catalogued by function , with genes grouped as receptors, signal transduction, and regulators.C. elegans protein sequences as queries reveals that the cyst nematodes possess homologs to many chemosensory genes. These include gpa‐3 and odr‐3 , tax‐2 and tax‐4 , and osm‐9 and egl‐4 (regulators). Indeed, RNAi knockdown of odr‐1 and odr‐3 receptors and tax‐2 and tax‐4 signal transducers resulted in defective chemotaxis to volatiles, non‐volatiles, and host root exudates in Meloidogyne incognita, supporting a role in host recognition for these chemosensory genes.mec‐10, mec‐4, trp‐1, and trp‐2 from C. elegans. Taken together, these genes represent excellent targets in cyst nematodes for RNAi to begin the unravelling of sensory mechanisms involved in host location/recognition by knockdown phenotype. Unlike C. elegans, where neuronal genes are mostly refractoryvia interference with cyst nematodes hatching or host location.Based on extensive studies of the chemosensory and mechanosensory mechanisms of 6Having a more comprehensive knowledge of the chemical constituents of root diffusates, especially the volatiles, will undoubtedly lead to our better understanding of the chemo‐ecological interactions between nematodes, their microbiome, and host plants. Clearly, there remains much to be established, in terms of the chemical signaling between cyst nematodes and their hosts. Chemical cues that play a role within rhizosphere also appear far more numerous and complex than we have so far anticipated and so, by fully characterizing the molecules involved in this chemical dialogue we will be better able to exploit this communication to disrupt specific signal pathways, especially when combined with host‐induced gene silencing approaches. This could be designed to either minimize attraction of host roots to nematodes or repel them. Alternatively, this approach can provide a mechanism to stimulate hatching in the absence of the host, or even to attract nematodes away from host roots, to non‐hosts or trap crops. Such environmentally sustainable approaches to the management of cyst nematodes, as well as other plant parasitic nematodes, offer enormous potential towards improving sustainable crop production. However, it is important that future efforts examine and exploit the chemical communication between soil organisms and plants and should take a multi‐disciplinary approach, including, amongst others breeding, biotechnology, plant pathology, as well as chemical ecology.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.Table S1. Biotic factors influencing cyst nematode hatchClick here for additional data file.


Prematurity has been identified as a risk factor for Autism Spectrum Disorder (ASD). The link between Autism Spectrum Disorder (ASD) and birth-week has not been strongly evidenced. We evaluated the correlation between the degree of prematurity and the incidence of autism in a cohort of 871 children born prematurely and followed from birth. The cohort was reduced to 416 premature infants born between 2011–2017 who were followed for 2–14 years, and analyzed according to birth week (degree of prematurity), and according to gender.43 children (10.3%) received a definite diagnosis of ASD. There was a significant correlation between birth week and the risk of ASD, with 22.6% of children diagnosed with ASD when born at 25 weeks, versus 6% of ASD diagnoses at 31 weeks of prematurity. For children born after 32 weeks, the incidence decreased to 8–12.5%. A strong link was found between earlier birth week and increased autism risk; the risk remained elevated during near-term prematurity in boys. A correlation between early birth week and an elevated risk for ASD was seen in all children, but accentuated in females, gradually decreasing as birth week progresses; in males the risk for ASD remains elevated for any birth week.A statistically significant increase in rates of autism was found with each additional week of prematurity. Females drove this direct risk related to degree of prematurity, while males had an elevated risk throughout prematurity weeks, even at near-term. We recommend including ASD screening in follow up of infants born prematurely, at all levels of prematurity. Born Too Soon: The Global Action Report on Preterm Birth" in 2012 reported that in 2010, approximately 1 in 10 babies was born prematurely, the incidence of prematurity did not change significantly in Israel (7.8–7.4% in 2015) Reviewers' comments:Reviewer's Responses to QuestionsComments to the Author1. Is the manuscript technically sound, and do the data support the conclusions?The manuscript must describe a technically sound piece of scientific research with data that supports the conclusions. Experiments must have been conducted rigorously, with appropriate controls, replication, and sample sizes. The conclusions must be drawn appropriately based on the data presented. Reviewer #1: NoReviewer #2: YesReviewer #3: Yes**********2. Has the statistical analysis been performed appropriately and rigorously? Reviewer #1: NoReviewer #2: YesReviewer #3: Yes**********3. Have the authors made all data underlying the findings in their manuscript fully available?PLOS Data policy requires authors to make all data underlying the findings described in their manuscript fully available without restriction, with rare exception (please refer to the Data Availability Statement in the manuscript PDF file). The data should be provided as part of the manuscript or its supporting information, or deposited to a public repository. For example, in addition to summary statistics, the data points behind means, medians and variance measures should be available. If there are restrictions on publicly sharing data—e.g. participant privacy or use of data from a third party—those must be specified.The Reviewer #1: NoReviewer #2: YesReviewer #3: No**********4. Is the manuscript presented in an intelligible fashion and written in standard English?PLOS ONE does not copyedit accepted manuscripts, so the language in submitted articles must be clear, correct, and unambiguous. Any typographical or grammatical errors should be corrected at revision, so please note any specific errors here.Reviewer #1: YesReviewer #2: YesReviewer #3: Yes**********5. Review Comments to the AuthorPlease use the space provided to explain your answers to the questions above. You may also include additional comments for the author, including concerns about dual publication, research ethics, or publication ethics. Reviewer #1: This study examines the association of ASD with preterm birth in 416 children from the ages of 2 to 11 years. More specifically, the authors seek to investigate the possibility of a gestational age gradient such that the risk of ASD decreases with each week of gestation from 24 to 37 weeks (from the lower boundary of extremely preterm to the upper boundary of later preterm) and how such a gradient might differ between males and females. They find that females have a significantly increased risk of ASD relative to males at lower gestational ages, whereas males maintain the same risk of ASD regardless of degree of prematurity.Comments1. The authors’ claim that such a gestational age gradient “has not been evidenced” in prior studies is inaccurate. At least two prior studies have demonstrated such a gradient (by week) with much larger epidemiological samples and with adjustment for a number of critical covariates and confounding variables that were not considered in the current study, presumably because the small sample size precluded their inclusion in statistical analyses. Please see Kuzniewicz et al., 2014 and Xie et al. (2017). In addition, Talmi et al. (2020) recently conducted a retrospective national cohort study of all children born in Israel from 2000-2012 using gestational age categories of extreme, very, moderate, and late preterm, showing a clear-cut gestational age gradient, as have a number of other large European population studies over the past decade, none of which are cited by the authors.2. The authors do not adequately address the loss to follow up of more than half of the sample (n=871) that was referred to them and the potential bias that this introduces into their results. At the very least, demographic and birth characteristics between those retained and those lost should be compared.3. Fetal growth restriction has been found to contribute, independently of gestational age, to ASD risk in a number of high-quality studies. The lack of consideration of FGR/SGA in this study is problematic, especially in that FGR may represent a distinct preterm phenotype that confers ASD risk by different mechanisms than GA.4. The large age range is somewhat problematic in that the preterm ASD phenotype is not well characterized. ASD classifications in the youngest children, whether positive or negative, may not be stable.5. It is now standard in autism epidemiological research to investigate ASD with and without intellectual disability (ID), as these are likely to differ to in etiology. The small sample size of the current study may have precluded meaningful statistical analysis addressing ID, but these data could have been presented for descriptive purposes. For example, one wonders how many of the EPT/VPT girls with ASD had ID and severe ID. The authors list a large number of cognitive measures, but they do not use these to describe their sample.6. The authors state that all ASD diagnoses were confirmed with the ADOS, but do not report any ADOS data, for example, the Modules used – which would have given a better sense of the level of function in the sample – or the severity scores for repetitive behaviors as well as social communicative impairment. These data are of interest in the characterization of any preterm ASD sample, given that the ADOS, a gold-standard measure, has rarely been used to characterize ASD in premies.7. The rationale for investigating the GA gradient for global development delay (GDD) and CP in addition to ASD is not clearly or compellingly explained. It is not clear if these are overlapping or mutually exclusive groupings, although the latter seems to be the case. In addition, it is not reported if any of the children classified as ASD had severe ID or severe gross motor impairment. This is problematic in that DSM-5 excludes an ASD diagnosis that is “better explained by intellectual disability  or global developmental delay”, and it is not uncommon for EPT-born children to have this severity of ID. In addition, severe motor impairment  could result in autism-like symptoms because of an inability to orient, direct attention, show, use gestures, etc.8. The authors do not adequately account for multiple birth in their statistical analyses. Approximately, 44% of the sample and 44% of those diagnosed with ASD were multiples, with the large majorities of those twins. (Information on monozygosity was not available.) T-tests and chi-squares are used to analyze differences between singletons and multiples in mean GA and distributions by gestational age, respectively, but the results are difficult to follow and the authors do not offer a clear interpretation. It would seem more effective to have included this variable as a covariate in the logistic regression analysis. I also wondered how many of those classified with ASD were from the same birth.9. It seems necessary to consider that the present findings are unreliable, given the small sample size and the inconsistency of the findings from a number of prior studies, which are not addressed in this report. It is notable, however, that others  have found that females are at relatively higher risk of ASD with ID vs. without ID than males, which may bear on the current findings, if ID were taken into account.Reviewer #2: This manuscript explores the relationship of birth week among premature infants to Autism Spectrum Disorder (ASD), to see if the degree of prematurity might predict the likelihood that ASD might develop. 416 children born from 2011 to 2017, from an original cohort of 871 premature children, followed for at least two years were studied retrospectively, looking at birth week and gender. All children were from a single child development referral center in Israel. Nearly half were twins or triplets, some of whom may have been monozygotic. The authors acknowledge that this may have affected the risk for developing ASD. A total of 43 children were felt to have ASD, and the risk was greatest for those who were more premature. While the overall risk of developing ASD is greater for males than females, an elevated risk of ASD with lower birth week was greatest for females, who had a higher risk than males for lower birth weeks and a lower risk than males for higher birth weeks. For males, the risk was relatively stable across the range of birth weeks.ASD is felt to be due to a combination of genetic and environmental factors, with brain insult and inflammation possibly contributing to the risk of developing it. Factors related to premature birth may contribute to the environmental components contributing to the development of ASD. Lower birth weight correlated with ASD risk in females, but not in males. A possible correlation between Cerebral Palsy (CP) and Global Developmental Delay (GDD) was also looked for. Measurements of ASD, CP and GDD by gender are shown across the range of birth weeks in the Figures 1-3. Table 1 summarizes the incidence of ASD by gender and birth week. A number of measures of cognitive function were utilized. Statistical analysis was appropriate.This study shows that prematurity is a risk factor for the development of ASD, and provides additional information regarding the relative risk attributable to birth week and gender. It is beyond the scope of this study to look at specific factors that might contribute to the risk of ASD in premature children, such as anoxic brain damage, intraventricular hemorrhage, presence or absence of Retinopathy of Prematurity (ROP), and treatment of Type 1 ROP with anti-VEGF injections. What can be said is that earlier intervention for those developing ASD may contribute to better outcomes, and recognition of the increased risk of ASD in premature children is therefore important. It is an interesting study.Reviewer #3: This is an interesting study with a retrospective look at prematurity and gender risk for ASD. There are several things that can strengthen it to make it more meaningful to autism researchers, clinician, and parents.Abstract: Children were followed for only “at least” two years, so the statement about girls should may need to be tempered to say that this relationship is evident only for girls whose symptoms are sufficiently apparent to be diagnosed by the age of 24 months, if that was the case (no info is given on when they were diagnosed). Girls with higher cognitive and language abilities are often not diagnosed until much later.Introduction: When the manuscript is revised, the CDC prevalence data needs to be updated to 1:54. “significant risk of developmental risk” may need to be revised – is developmental delay the intended meaning?Please clarify when the children were evaluated. It is stated that you only included children seen in the clinic up to the age of 2 to prevent loss to follow up, but several of the cognitive measures listed are for older children only. At what ages where the cognitive evaluations completed? It is apparent from the age range (2-14.2 years) that cognitive testing may have occurred at any point. This is a retrospective study, so presumably cognitive testing was done for a reason unrelated to the study, which should be addressed at least briefly to determine possibility of bias in the sample. For example, if the cognitive testing was performed as a matter of routine follow up for a premature child, that is less biased than if they were only performed when disability was suspected, introducing at least some selection bias in the sample. This issue is also related to the comment above about females across the spectrum who are typically not diagnosed until later in life.How old were the children when they were diagnosed? That is an important proxy for interpretation of the severity of the ASD.Also, it would be helpful to know co-morbid diagnoses. Co-morbidity data would also aid in the interpretation of the findings. Are the females diagnosed with ASD in the lower gestational ages also affected by multiple disabilities or is ASD the primary risk? All of this information is more helpful than a blanket statement about risk and females as if they are all equivalent in terms of ASD. The current data are helpful, but MORE data from the records would significantly improve the interpretation and application of the study. Since Cognitive testing was completed, that is one example of data that aids in interpretation. Females with Intellectual Disability (ID) present very differently than females without ID, and may or may not have been identified yet to be included in the ASD data.Line 204 -- What is the reference for “1:5 male/ female ratio of idiopathic autism”? And is this stated correctly? This sentence is used following presentation of the data of 20 males with ASD in the more advanced pregnancies compared to 4 females.Line 265 – typo “especially form the less concerning group of patients born,” there are other suspected typos throughout. Discussion: The discussion somewhat buries the lead in the statement (line 254) “However, after 33 weeks the incidence remains elevated and does not continue to decrease, following the pattern of increased risk in males, in the same gender distribution as "idiopathic" autism (33)”Following that, it is surprisingly light on discussion of females with ASD, which is a very current topic in the autism community right now. Since the discussion acknowledges that the finding of sustained risk for males not only in prematurity but throughout gestation time is not novel, I think more discussion about the meaning of the finding for females is expected here, as that is the novel finding and the title of the article. Much is made of the findings for multiple births, which is also less than novel.**********what does this mean?). If published, this will include your full peer review and any attached files.6. PLOS authors have the option to publish the peer review history of their article  digital diagnostic tool,  24 May 2020Here are our specific corrections and comments: # Academic editorPONE-D-19-35927Autism risk linked to prematurity is more accentuated in girlsPLOS ONEJournal requirements:When submitting your revision, we need you to address these additional requirements:http://www.plosone.org/attachments/PLOSOne_formatting_sample_main_body.pdf and http://www.plosone.org/attachments/PLOSOne_formatting_sample_title_authors_affiliations.pdf1. Please ensure that your manuscript meets PLOS ONE's style requirements, including those for file naming. The PLOS ONE style templates can be found at 2. We note that you have stated that you will provide repository information for your data at acceptance. Should your manuscript be accepted for publication, we will hold it until you provide the relevant accession numbers or DOIs necessary to access your data. If you wish to make changes to your Data Availability statement, please describe these changes in your cover letter and we will update your Data Availability statement to reflect the information you provide.Author response: Ok, Thank you.[Note: HTML markup is below. Please do not edit.]Reviewers' comments:Reviewer's Responses to Questions#Comments to the Author1. Is the manuscript technically sound, and do the data support the conclusions?The manuscript must describe a technically sound piece of scientific research with data that supports the conclusions. Experiments must have been conducted rigorously, with appropriate controls, replication, and sample sizes. The conclusions must be drawn appropriately based on the data presented.Reviewer #1: NoReviewer #2: YesReviewer #3: YesAuthor response:We thank the reviewers, as per the comments of the reviewers, changes were made to the article. 2. Has the statistical analysis been performed appropriately and rigorously?Reviewer #1: NoReviewer #2: YesReviewer #3: Yes Author response: In accordance with the comments of reviewer 1, changes in the statistical analyses were made in the article3. Have the authors made all data underlying the findings in their manuscript fully available?The PLOS Data policy requires authors to make all data underlying the findings described in their manuscript fully available without restriction, with rare exception (please refer to the Data Availability Statement in the manuscript PDF file). The data should be provided as part of the manuscript or its supporting information, or deposited to a public repository. For example, in addition to summary statistics, the data points behind means, medians and variance measures should be available. If there are restrictions on publicly sharing data—e.g. participant privacy or use of data from a third party—those must be specified.Reviewer #1: NoReviewer #2: YesReviewer #3: NoAuthor response:In accordance with the comments, we now allow access to all anonymous research data, hence we will need to translate the information to English. 4. Is the manuscript presented in an intelligible fashion and written in standard English?PLOS ONE does not copyedit accepted manuscripts, so the language in submitted articles must be clear, correct, and unambiguous. Any typographical or grammatical errors should be corrected at revision, so please note any specific errors here.Reviewer #1: YesReviewer #2: YesReviewer #3: Yes Author response:Thanks.5. Review Comments to the AuthorPlease use the space provided to explain your answers to the questions above. You may also include additional comments for the author, including concerns about dual publication, research ethics, or publication ethics. Reviewer #1Reviewer #1: This study examines the association of ASD with preterm birth in 416 children from the ages of 2 to 11 years. More specifically, the authors seek to investigate the possibility of a gestational age gradient such that the risk of ASD decreases with each week of gestation from 24 to 37 weeks (from the lower boundary of extremely preterm to the upper boundary of later preterm) and how such a gradient might differ between males and females. They find that females have a significantly increased risk of ASD relative to males at lower gestational ages, whereas males maintain the same risk of ASD regardless of degree of prematurity.Comments1. The authors’ claim that such a gestational age gradient “has not been evidenced” in prior studies is inaccurate. At least two prior studies have demonstrated such a gradient (by week) with much larger epidemiological samples and with adjustment for a number of critical covariates and confounding variables that were not considered in the current study, presumably because the small sample size precluded their inclusion in statistical analyses. Please see Kuzniewicz et al., 2014 and Xie et al. (2017). In addition, Talmi et al. (2020) recently conducted a retrospective national cohort study of all children born in Israel from 2000-2012 using gestational age categories of extreme, very, moderate, and late preterm, showing a clear-cut gestational age gradient, as have a number of other large European population studies over the past decade, none of which are cited by the authors.Author response:We thank the reviewer for the constructive comments. We now addressed the aforementioned studies in our introduction. The analysis in the studies was categorical and divided prematurity in three large groups based on three levels of prematurity (uneven groups in terms of weeks- comparing two to six weeks clumped together). Other studies mentioned by the reviewer analyze birth weight and not gestational week. Our study offers risk data per each gestational week as a continuum and emphasizes the gender differences, which were not demonstrated in prior studies. 2. The authors do not adequately address the loss to follow up of more than half of the sample (n=871) that was referred to them and the potential bias that this introduces into their results. At the very least, demographic and birth characteristics between those retained and those lost should be compared.Author response:Thank you, we added demographic and birth characteristics between those retained and those lost: "Of the 838 children identified who were born premature, 416 children were seen after the age of two years; 422 children were excluded. Of the excluded 229 (54.3%) were male, and 193 (45.7%) were female. The average birth week was 32.3 weeks (SD=2.5) and the average birth weight was 1670 grams (SD=486), which is significantly higher compared to children who were seen after the age of two years, t(834)=7.26, p<0.001, t(819)=6.65, p<0.001, respectively". 3. Fetal growth restriction has been found to contribute, independently of gestational age, to ASD risk in a number of high-quality studies. The lack of consideration of FGR/SGA in this study is problematic, especially in that FGR may represent a distinct preterm phenotype that confers ASD risk by different mechanisms than GA.Author response: Thank you for your comments, we re-analyzed the data. Effect for IUGR was not significant and the conclusions remained the same. We added:"We also examined IUGR/SGA. Thirty seven children had IUGR/SGA, 20 were males (54%) and 17 were females (46%). Birth week ranged from week 24-37, with an average of 30.8 weeks (SD= 3.9). Birth weight ranged from 580-2200 grams, with an average of 1130 grams (SD=448).No correlation was found between IUGR/SGA and ASD, Ҳ² (1)=0.2, p>.05., Thirty four of IUGR without ASD (92%) and 3 IUGR with ASD (8%). In order to examine whether IUGR had an effect on the results we exclude these 37 children from model. The model was significant Ҳ²(5)=17.74, p<0.01, effect for IUGR was not significant Ҳ²(1)=0, p>0.05 and the conclusions remained the same: the probability to ASD decreased for females with each birth week progress while for males it remained relatively constant".4. The large age range is somewhat problematic in that the preterm ASD phenotype is not well characterized. ASD classifications in the youngest children, whether positive or negative, may not be stable.Author response:Thank you for affirming our approach, this is indeed the reason we included only the children that were in the 416 cohort, with confirmed ASD after age 2 years, in the model and multiple comparisons.5. It is now standard in autism epidemiological research to investigate ASD with and without intellectual disability (ID), as these are likely to differ to in etiology. The small sample size of the current study may have precluded meaningful statistical analysis addressing ID, but these data could have been presented for descriptive purposes. For example, one wonders how many of the EPT/VPT girls with ASD had ID and severe ID. The authors list a large number of cognitive measures, but they do not use these to describe their sample.Author response:Thank you for this important comment, the requested analysis strengthened significantly our model and revealed an important finding: Indeed, we have followed your indication and examined the effect children with ID might have on the model, but found no contribution of this factor (ID) to the model. Moreover, children with ID were included in the group of 37 children with IUGR. Thus, by excluding them all in the model without IUGR and obtaining similar results, we confirm that no added effect is measured there. Since the study group is young, we also added a model including GDD comorbidity (as a precursor of ID):"68 children (16.3%) received a diagnosis of GDD, thirteen of them have ASD (19%). Adding the GDD as an intervening factor to the model, strengthened the model slightly Ҳ²(6)=22.81, p<0.001 and the GDD is significant Ҳ²(1)=6.48, p<0.05 as well as its interaction with birth week and gender Ҳ²(1)=4.22, p<0.05. When constructing a graph based on the nominal logistic model prediction, we see that while for females trends remain similar, the proportion of ASD drops as birth week increases. For females with GDD, the same trend is kept but with a higher propensity towards ASD. For males however, trends reverse. While for males without GDD the trend is relatively constant across birth weeks, for males with GDD the proportion of males with ASD increases with birth week ".6. The authors state that all ASD diagnoses were confirmed with the ADOS, but do not report any ADOS data, for example, the Modules used – which would have given a better sense of the level of function in the sample – or the severity scores for repetitive behaviors as well as social communicative impairment. These data are of interest in the characterization of any preterm ASD sample, given that the ADOS, a gold-standard measure, has rarely been used to characterize ASD in premies.Author response:Thank you, the ADOS data was not included since the sample comprised of young children. For most of the participants, Module 1 was used, more recently the Toddler adaptation for Module 1. Because of the age group and comorbidities related to prematurity, we assumed that severity data will not be conclusive. Based on the remarks, we are in the process of expanding the study and examining further the clinical symptoms of ASD in our premature group in comparison to a term group, but the results are beyond the scope of this paper. 7. The rationale for investigating the GA gradient for global development delay (GDD) and CP in addition to ASD is not clearly or compellingly explained. It is not clear if these are overlapping or mutually exclusive groupings, although the latter seems to be the case. In addition, it is not reported if any of the children classified as ASD had severe ID or severe gross motor impairment. This is problematic in that DSM-5 excludes an ASD diagnosis that is “better explained by intellectual disability  or global developmental delay”, and it is not uncommon for EPT-born children to have this severity of ID. In addition, severe motor impairment  could result in autism-like symptoms because of an inability to orient, direct attention, show, use gestures, etc.Author response:Thank you for your useful comments, we added:"68 children (16.3%) received a diagnosis of GDD, thirteen of them have ASD (19%). Adding the GDD as an intervening factor to the model, strengthened the model slightly Ҳ²(6)=22.81, p<0.001 and the GDD is significant Ҳ²(1)=6.48, p<0.05 as well as its interaction with birth week and gender Ҳ²(1)=4.22, p<0.05. When constructing a graph based on the nominal logistic model prediction, we see that while for females trends remain similar, the proportion of ASD drops as birth week increases. For females with GDD, the same trend is kept but with a higher propensity towards ASD. For males however, trends reverse. While for males without GDD the trend is relatively constant across birth weeks, for males with GDD the proportion of males with ASD increases with birth week ".In regards to children with an additional severe motor diagnosis we included it as a formal diagnosis of cerebral palsy. It is true that the clinical symptoms of severe CP or of global developmental delay may overlap with autism symptoms, but this is the main clinical expertise of our center- to be able to differentiate between CP or GDD as a main diagnosis verses CP or GDD as a comorbidity to an ASD diagnosis. Our diagnostic team is comprised not only of neurodevelopmental doctors and psychologists, but also occupational, language and physical therapists. We assess all children by formal assessments for all developmental diagnoses enabling specific and accurate differential diagnosis. Description of our assessment of CP was published in Gabis et al., 2015.8. The authors do not adequately account for multiple birth in their statistical analyses. Approximately, 44% of the sample and 44% of those diagnosed with ASD were multiples, with the large majorities of those twins. (Information on monozygosity was not available.) T-tests and chi-squares are used to analyze differences between singletons and multiples in mean GA and distributions by gestational age, respectively, but the results are difficult to follow and the authors do not offer a clear interpretation. It would seem more effective to have included this variable as a covariate in the logistic regression analysis. I also wondered how many of those classified with ASD were from the same birth.Author response:Thank you, we added:"Out of 184 multiple pregnancies 19 had ASD (10.3%). We examined the distribution of the proportion of multiple pregnancies with ASD across birth weeks but no particular significance was found between earlier and later birth weeks, nor were we able to construct a significant model indicating twins as a significant intervening factor Ҳ² (1)=1.52, p>.05."9. It seems necessary to consider that the present findings are unreliable, given the small sample size and the inconsistency of the findings from a number of prior studies, which are not addressed in this report. It is notable, however, that others  have found that females are at relatively higher risk of ASD with ID vs. without ID than males, which may bear on the current findings, if ID were taken into account.Author response:We thank the reviewer for the important comment. Indeed, similar findings suggestive of a relationship between prematurity and autism were found in other older studies , and also newer studies . We discussed limitations and added value of our study in the introduction as referring to the above comment. Reviewer #2Reviewer #2: This manuscript explores the relationship of birth week among premature infants to Autism Spectrum Disorder (ASD), to see if the degree of prematurity might predict the likelihood that ASD might develop. 416 children born from 2011 to 2017, from an original cohort of 871 premature children, followed for at least two years were studied retrospectively, looking at birth week and gender. All children were from a single child development referral center in Israel. Nearly half were twins or triplets, some of whom may have been monozygotic. The authors acknowledge that this may have affected the risk for developing ASD. A total of 43 children were felt to have ASD, and the risk was greatest for those who were more premature. While the overall risk of developing ASD is greater for males than females, an elevated risk of ASD with lower birth week was greatest for females, who had a higher risk than males for lower birth weeks and a lower risk than males for higher birth weeks. For males, the risk was relatively stable across the range of birth weeks.ASD is felt to be due to a combination of genetic and environmental factors, with brain insult and inflammation possibly contributing to the risk of developing it. Factors related to premature birth may contribute to the environmental components contributing to the development of ASD. Lower birth weight correlated with ASD risk in females, but not in males. A possible correlation between Cerebral Palsy (CP) and Global Developmental Delay (GDD) was also looked for. Measurements of ASD, CP and GDD by gender are shown across the range of birth weeks in the Figures 1-3. Table 1 summarizes the incidence of ASD by gender and birth week. A number of measures of cognitive function were utilized. Statistical analysis was appropriate.This study shows that prematurity is a risk factor for the development of ASD, and provides additional information regarding the relative risk attributable to birth week and gender. It is beyond the scope of this study to look at specific factors that might contribute to the risk of ASD in premature children, such as anoxic brain damage, intraventricular hemorrhage, presence or absence of Retinopathy of Prematurity (ROP), and treatment of Type 1 ROP with anti-VEGF injections. What can be said is that earlier intervention for those developing ASD may contribute to better outcomes, and recognition of the increased risk of ASD in premature children is therefore important. It is an interesting study.Author response:Thank youReviewer #3Reviewer #3: This is an interesting study with a retrospective look at prematurity and gender risk for ASD. There are several things that can strengthen it to make it more meaningful to autism researchers, clinician, and parents.Abstract: Children were followed for only “at least” two years, so the statement about girls should may need to be tempered to say that this relationship is evident only for girls whose symptoms are sufficiently apparent to be diagnosed by the age of 24 months, if that was the case (no info is given on when they were diagnosed). Girls with higher cognitive and language abilities are often not diagnosed until much later.Author response:We agree, for this reason we limited the group to two years and above. Some of the girls were indeed diagnosed later, and as such included in the analysis. However we did not find a significant age difference in the age of diagnosis. Introduction: When the manuscript is revised, the CDC prevalence data needs to be updated to 1:54. “significant risk of developmental risk” may need to be revised – is developmental delay the intended meaning? Author response:Thank you for your attention, corrected. Please clarify when the children were evaluated. It is stated that you only included children seen in the clinic up to the age of 2 to prevent loss to follow up, but several of the cognitive measures listed are for older children only. At what ages where the cognitive evaluations completed? It is apparent from the age range (2-14.2 years) that cognitive testing may have occurred at any point. This is a retrospective study, so presumably cognitive testing was done for a reason unrelated to the study, which should be addressed at least briefly to determine possibility of bias in the sample. For example, if the cognitive testing was performed as a matter of routine follow up for a premature child, that is less biased than if they were only performed when disability was suspected, introducing at least some selection bias in the sample. This issue is also related to the comment above about females across the spectrum who are typically not diagnosed until later in life.Author response:Thank you for your remark. The cognitive evaluation is performed initially during the initial diagnostic process and, according to the DSM-5 and MOH guidelines, is mandatory when a diagnosis of ASD made in order to establish that ID is not the main diagnosis (may accompany ASD). As such, all cognitive assessments were performed at the initial evaluation of autism and the instruments used are Mullen (up to age three) and IQ tests, such as WPPSI-III (until school age) and WISC-R (school age). How old were the children when they were diagnosed? That is an important proxy for interpretation of the severity of the ASD.Author response:Thank you for your remark, we added age of diagnosis:"The age of initial ASD diagnosis was in the range of 1.1-8.5 years, with mean age of 3.5 years (SD=1.8)". Also, it would be helpful to know co-morbid diagnoses. Co-morbidity data would also aid in the interpretation of the findings. Are the females diagnosed with ASD in the lower gestational ages also affected by multiple disabilities or is ASD the primary risk? All of this information is more helpful than a blanket statement about risk and females as if they are all equivalent in terms of ASD. The current data are helpful, but MORE data from the records would significantly improve the interpretation and application of the study. Since Cognitive testing was completed, that is one example of data that aids in interpretation. Females with Intellectual Disability (ID) present very differently than females without ID, and may or may not have been identified yet to be included in the ASD data.Author response:Thank you, we examined the effect ID children might have on the model and each pair comparisons for significance but found no contribution of this factor (ID) to the model. Moreover, children with ID were all included in the group of 37 children with IUGR. Thus, by excluding them all in the model without IUGR children and obtaining similar results we confirm that no added effect is measured there. We also added a model with GDD:"68 children (16.3%) received a diagnosis of GDD, thirteen of them have ASD (19%). Adding the GDD as an intervening factor to the model strengthened the model slightly Ҳ²(6)=22.81, p<0.001 and the GDD is significant Ҳ²(1)=6.48, p<0.05 as well as its interaction with birth week and gender Ҳ²(1)=4.22, p<0.05. When constructing a graph based on the nominal logistic model prediction, we see that while for females trends remain similar; the proportion of ASD drops as birth week increases, for females with GDD the same trend is kept but with a higher propensity to ASD. For males however, trends reverse; while for non GDD males the trend is relatively constant across birth weeks, for males with GDD the proportion of ASD males increases with birth week ".Line 204 -- What is the reference for “1:5 male/ female ratio of idiopathic autism”? And is this stated correctly? This sentence is used following presentation of the data of 20 males with ASD in the more advanced pregnancies compared to 4 females.Author response:Thank you, we added reference: Gabis L V., Pomeroy J. An etiologic classification of autism spectrum disorders. Isr Med Assoc J. 2014;16(5):295–8.Line 265 – typo “especially form the less concerning group of patients born,” there are other suspected typos throughout. Author response:We thank the reviewer for turning our attention to the typos. Based on the reviewer’s note we made the change to line 273, 397, and throughout the article.Discussion: The discussion somewhat buries the lead in the statement (line 254) “However, after 33 weeks the incidence remains elevated and does not continue to decrease, following the pattern of increased risk in males, in the same gender distribution as "idiopathic" autism (33)”Following that, it is surprisingly light on discussion of females with ASD, which is a very current topic in the autism community right now. Since the discussion acknowledges that the finding of sustained risk for males not only in prematurity but throughout gestation time is not novel, I think more discussion about the meaning of the finding for females is expected here, as that is the novel finding and the title of the article. Much is made of the findings for multiple births, which is also less than novel.Author response:Thank you very much for the opportunity to include our insights on gender differential vulnerability in the developing brain. We added a paragraph and new references in the discussion. 6. PLOS authors have the option to publish the peer review history of their article (what does this mean?). If published, this will include your full peer review and any attached files.If you choose “no”, your identity will remain anonymous but your review may still be made public.Author response: OK. 9 Jul 2020PONE-D-19-35927R1Autism risk linked to prematurity is more accentuated in girlsPLOS ONEDear Dr. Gabis,Thank you for submitting your revised manuscript to PLOS ONE. After careful consideration, there are still a few minor issues that will need to be addressed. In particular, one of the reviewers made the point that attrition could have affected your estimates. Please address this statistically or by noting in the text how this may impact your results and interpretations. The reviewer also noted a typo. Please proof your manuscript carefully as PLOS ONE will not copy edit your manuscript for typographical errors. plosone@plos.org. When you're ready to submit your revision, log on to https://www.editorialmanager.com/pone/ and select the 'Submissions Needing Revision' folder to locate your manuscript file.Please submit your revised manuscript by Aug 23 2020 11:59PM. If you will need more time than this to complete your revisions, please reply to this message or contact the journal office at Please include the following items when submitting your revised manuscript:A rebuttal letter that responds to each point raised by the academic editor and reviewer(s). You should upload this letter as a separate file labeled 'Response to Reviewers'.A marked-up copy of your manuscript that highlights changes made to the original version. You should upload this as a separate file labeled 'Revised Manuscript with Track Changes'.An unmarked version of your revised paper without tracked changes. You should upload this as a separate file labeled 'Manuscript'.If you would like to make changes to your financial disclosure, please include your updated statement in your cover letter. Guidelines for resubmitting your figure files are available below the reviewer comments at the end of this letter.http://journals.plos.org/plosone/s/submission-guidelines#loc-laboratory-protocolsIf applicable, we recommend that you deposit your laboratory protocols in protocols.io to enhance the reproducibility of your results. Protocols.io assigns your protocol its own identifier (DOI) so that it can be cited independently in the future. For instructions see: We look forward to receiving your revised manuscript.Kind regards,Eric J. Moody, Ph.D.Academic EditorPLOS ONE[Note: HTML markup is below. Please do not edit.]Reviewers' comments:Reviewer's Responses to QuestionsComments to the Author1. If the authors have adequately addressed your comments raised in a previous round of review and you feel that this manuscript is now acceptable for publication, you may indicate that here to bypass the “Comments to the Author” section, enter your conflict of interest statement in the “Confidential to Editor” section, and submit your "Accept" recommendation.Reviewer #2: All comments have been addressedReviewer #4: (No Response)**********2. Is the manuscript technically sound, and do the data support the conclusions?The manuscript must describe a technically sound piece of scientific research with data that supports the conclusions. Experiments must have been conducted rigorously, with appropriate controls, replication, and sample sizes. The conclusions must be drawn appropriately based on the data presented. Reviewer #2: YesReviewer #4: Yes**********3. Has the statistical analysis been performed appropriately and rigorously? Reviewer #2: YesReviewer #4: Yes**********4. Have the authors made all data underlying the findings in their manuscript fully available?PLOS Data policy requires authors to make all data underlying the findings described in their manuscript fully available without restriction, with rare exception (please refer to the Data Availability Statement in the manuscript PDF file). The data should be provided as part of the manuscript or its supporting information, or deposited to a public repository. For example, in addition to summary statistics, the data points behind means, medians and variance measures should be available. If there are restrictions on publicly sharing data—e.g. participant privacy or use of data from a third party—those must be specified.The Reviewer #2: YesReviewer #4: No**********5. Is the manuscript presented in an intelligible fashion and written in standard English?PLOS ONE does not copyedit accepted manuscripts, so the language in submitted articles must be clear, correct, and unambiguous. Any typographical or grammatical errors should be corrected at revision, so please note any specific errors here.Reviewer #2: YesReviewer #4: Yes**********6. Review Comments to the AuthorPlease use the space provided to explain your answers to the questions above. You may also include additional comments for the author, including concerns about dual publication, research ethics, or publication ethics. Reviewer #2: I believe that the questons posed by the three reviewers have been adequately answered. I have nothing further to add, apart from my initial comments.Reviewer #4: line 287: the prevalence of ASD in the study sample could have been inflated by the high attrition rate (the infants without ASD are more likely to not coming back to the clinic.)line 419: typo**********what does this mean?). If published, this will include your full peer review and any attached files.7. PLOS authors have the option to publish the peer review history of their article  digital diagnostic tool,  16 Jul 2020# Academic editorPONE-D-19-35927Autism risk linked to prematurity is more accentuated in girlsPLOS ONE6. Review Comments to the Author#Comments to the AuthorReviewer #2: I believe that the questions posed by the three reviewers have been adequately answered. I have nothing further to add, apart from my initial comments.Author response: Thank you. Reviewer #4: line 287: the prevalence of ASD in the study sample could have been inflated by the high attrition rate (the infants without ASD are more likely to not coming back to the clinic.)Author response: Thank you for this very important remark and we will try to address it in different ways: - In regards to developmental follow up of non –ASD children, loss to follow up after the age of two years may be suspected for premature children with typical development, which may influence the rate of disability in the cohort. However, this may influence incidence of all developmental issues and not only ASD, since a high percentage of children born prematurely may have global developmental delay or delays in a single area of development.- This matter of declining numbers in the control cohort needs to be addressed, and is in the first steps of data exploration, as explained in the abstract in rows 29-30. In order to analyze this possible bias we recalculated the incidence for the whole cohort of 817 children followed from birth, including infants that did not reach two years. The same model and multiple comparisons were implemented and still showed similar significances as in the subset cohort of 416 children with a validated diagnosis of positive/ negative ASD. Within the 391 that were removed from the main analysis cohort of 416 children, there were 389 children with no ASD diagnosis, and 2 females born week 27 and 36 (*10 exceptions excluded) with ASD. With this additional analysis, the principal conclusion was even strengthened- that the increased incidence with lower birth week is driven mainly by girls. So to answer the reviewer’s comment, an increase in controls does not seem to affect the conclusion and vice versa.- line 419: typoAuthor response: Corrected, thank you.Attachmentresponse to reviewers PLOS ONE.docxSubmitted filename: Click here for additional data file. 20 Jul 2020Autism risk linked to prematurity is more accentuated in girlsPONE-D-19-35927R2Dear Dr. Gabis,We’re pleased to inform you that your manuscript has been judged scientifically suitable for publication and will be formally accepted for publication once it meets all outstanding technical requirements.Within one week, you’ll receive an e-mail detailing the required amendments. When these have been addressed, you’ll receive a formal acceptance letter and your manuscript will be scheduled for publication.http://www.editorialmanager.com/pone/, click the 'Update My Information' link at the top of the page, and double check that your user information is up-to-date. If you have any billing related questions, please contact our Author Billing department directly at authorbilling@plos.org.An invoice for payment will follow shortly after the formal acceptance. To ensure an efficient process, please log into Editorial Manager at onepress@plos.org.If your institution or institutions have a press office, please notify them about your upcoming paper to help maximize its impact. If they’ll be preparing press materials, please inform our press team as soon as possible -- no later than 48 hours after receiving the formal acceptance. Your manuscript will remain under strict press embargo until 2 pm Eastern Time on the date of publication. For more information, please contact Kind regards,Eric J. Moody, Ph.D.Academic EditorPLOS ONEAdditional Editor Comments :Reviewers' comments: 11 Aug 2020PONE-D-19-35927R2 Autism risk linked to prematurity is more accentuated in girls Dear Dr. Gabis:I'm pleased to inform you that your manuscript has been deemed suitable for publication in PLOS ONE. Congratulations! Your manuscript is now with our production department. onepress@plos.org.If your institution or institutions have a press office, please let them know about your upcoming paper now to help maximize its impact. If they'll be preparing press materials, please inform our press team within the next 48 hours. Your manuscript will remain under strict press embargo until 2 pm Eastern Time on the date of publication. For more information please contact plosone@plos.org. If we can help with anything else, please email us at Thank you for submitting your work to PLOS ONE and supporting open access. Kind regards, PLOS ONE Editorial Office Staffon behalf ofDr. Eric J. Moody Academic EditorPLOS ONE


Short birth interval, defined as a birth-to-birth interval less than 33 months, is associated with adverse maternal and child outcomes. Evidence regarding the association of maternal socioeconomic status and short birth interval is inconclusive. Factors contributing to the socioeconomic inequality of short birth interval have also not been investigated. The current study assessed socioeconomic inequality in short birth interval and its contributing factors in Ethiopia.Data from 8448 women collected in the 2016 Ethiopia Demographic and Health survey were included in the study. Socioeconomic inequality in short birth interval was the outcome variable. Erreygers normalized concentration index (ECI) and concentration curves were used to measure and illustrate socioeconomic-related inequality in short birth interval, respectively. Decomposition analysis was performed to identify factors explaining the socioeconomic-related inequality in short birth interval.P < 0.0001); indicating that short birth interval was more concentrated among the poor. Decomposition analysis indicated that wealth quintiles (74.2%), administrative regions (26.4%), and not listening to the radio (5.6%) were the major contributors to the pro-poor socioeconomic inequalities in short birth interval.The Erreygers normalized concentration index for short birth interval was − 0.0478 (SE = 0.0062) and differed significantly from zero (There was a pro-poor inequality of short birth interval in Ethiopia. Strengthening the implementation of poverty alleviation programs may improve the population’s socioeconomic status and reduce the associated inequality in short birth interval. Health inequality is defined as a systematic difference in health across individuals or according to socially relevant groupings such as between more and less advantaged groups –3. GlobaAlthough reducing inequality is the current central objective of health policy in many countries including Ethiopia, progress has been inadequate , 17. StuPrevious studies have documented the adverse maternal and child health outcomes associated with short birth interval, defined as inter-birth interval of less than 33 months , such asUnlike the aforementioned previous studies –33 perfoThe existing empirical studies –33 estimA simple but comprehensive estimate of the socioeconomic inequality in short birth interval and its contributors is needed to inform policy makers and public health planners for targeted action. In this regard, the CI and decomposition analyses , 38 wereThe data for this study were extracted from the 2016 Ethiopia Demographic and Health Survey (EDHS). The 2016 EDHS is the fourth and most recent nationally representative Demographic and Health Survey (DHS) in the country. This cross-sectional survey is carried out every 5 years to provide health and health-related indicators for the country as a whole, for urban and rural areas separately, and for each of the nine regions and the two administrative cities. The survey employed a two-stage stratified cluster sampling method. The detailed sampling procedure is presented in the full EDHS report . A totalSocioeconomic inequality in short birth interval was the outcome variable in this study. Birth interval was defined as the time between the birth of the child under study (index child) and the immediately preceding birth . In the The socioeconomic inequality of short birth interval can be expressed as the covariance between short birth interval and the fractional rank in the living standards distribution . It was then classified into either pro-poor or pro-rich or no inequality problem. Detailed explanation about estimating the socioeconomic inequality of short birth interval can be found in the data analysis section of this paper.Key explanatory variables included in the decomposition analysis were selected after reviewing relevant literature , 49, 50.Wealth index was used as a measure of the socioeconomic status of the household. Households were given scores based on the number and kinds of consumer goods they own, ranging from a television to a bicycle or car, in addition to housing characteristics such as type of drinking water source, flooring materials, toilet and sanitation facilities. A weight or factor score generated through principal components analysis was given to each household asset which was believed to be indicative of the wealth status and for which data was collected. Then standardization of the resulting asset scores to a standard normal distribution with a mean of zero and a standard deviation of one was made. A standardized score for each asset, which varies depending on the household possession status of the asset, was assigned for each household. The scores were summed by the households and the total score of the household was used to rank the individuals. The sample was then divided into population quintiles, that is, five groups, each comprising 20% of the population. The lowest 20% quintile was assigned to the poorest households, the next 20% quintile to the poor households, followed by another 20% quintile for the middle-class households, and finally the top 40% quintile for the rich and richest households. Literature has documented that wealth quintile represents a more long-term (permanent) economic status than either income or consumption does and it is also much easier to implement , 52. FurP-values were computed using Pearson’s chi-squared test. Sampling weight was used throughout the analyses  to adjust for the non-proportional allocation of the sample to different regions, to their urban and rural areas, and the possible differences in response rates. It ensures the actual representativeness of the survey results at both the national and domain levels. Further explanation of the weighting procedure can be found in the EDHS report [Descriptive statistics, the frequency with percent, were used to illustrate the distribution of respondents’ background characteristics. S report .A concentration index (CI) was computed to measure the socioeconomic inequality in short birth interval. The concentration index is a relative measure of inequality and is defined as twice the area between the concentration curve and the line of equality (the 45-degree line) , h represents health variable , and r is the cumulative percentage that each woman represents over the total population after ranking short birth interval by the wealth index. For an unbound variable, the index ranges between − 1 and 1 while it ranges from μ − 1 to 1 − μ for bounded variables [Where c represents concentration index, ariables . Multiplariables .ECI is Erreygers concentration index, CI(y) is the generalized concentration index and μ is the mean of the health variable .To consider the bounded nature of binary health variables, such as short birth interval, Erreygers  proposedConcentration curves were used to graphically depict the socioeconomic related inequality in short birth interval. The curves demonstrate the cumulative percentage of short birth interval on the y-axis against the cumulative percentage of women ranked by the wealth index on the x-axis, sorted from the poorest to the richest. In the case when there is no socioeconomic-related inequality (perfect equity), the concentration index is zero. In other words, if everyone, irrespective of her living standards, has exactly the same value of the short birth interval, the concentration curve lies at a 45-degree line, running from the bottom left-hand corner to the top right-hand corner. This is known as the line of equality , this inspection is not sufficient to conclude whether or not the concentration curve dominance is statistically significant. Therefore, a dominance test was performed to examine the statistical significance of the difference between the concentration curve and the line of equality . It helps to identify significance of difference between ordinates of curves at a number of quantiles. To achieve this, a multiple comparison approach that considered multiple testing using critical values from the Studentized Maximum Modulus (SMM) distribution was performed. If a critical value for 5% significance level to test significance between co-ordinates is used, then as the number of points of comparison is increased, the more likely it becomes that the null hypothesis could be falsely rejected. In contrast, if too few points of comparison were considered, dominance across the full range of distribution cannot be tested. Hence, it is necessary to correct the critical value for the number of comparisons made. The SSM distribution takes account of the number of points of comparison. Thus, 19 equally spaced quantile points, a recommended points choice, and ere used . Furtherere used .Decomposition of the concentration index was also performed to identify the relative contribution of various factors to the socioeconomic-related inequality in short birth interval , 34, 40.y is a health outcome variable , xk is a set of socioeconomic determinants of health outcome, α is an intercept, βk is the coefficient of xk, μ is the mean of y, xk, Ck is the concentration index for xk , and GCε denotes the generalized concentration index for the error term/residual term (ε), ηk) of y  with respect to ε) reflects the inequality in short birth interval that cannot be explained by systematic variations across income groups in the xk, which should approach zero for a well specified model [Where interval , 40. Theed model , 40. Samed model , 57, 58. was used for the statistical analysis.Stata version 14 A total of 8448 women were included in the analysis. The prevalence of short birth interval in Ethiopia was 45.8% (95% CI: 42.91–48.62).Table P < 0.0001). This indicates that short birth interval was disproportionately concentrated among the poor. Figure p-value < 0.001).The weighted Erreygers normalized concentration index (ECI) for short birth interval was − 0.0478 (SE = 0.0062) and differed significantly from zero , as shown by the adjusted percentage contribution of inequality. The regions where women lived, in general, were responsible for 26.4% of the socioeconomic inequality. Not listening to radio also contributed to the socioeconomic inequalities in short birth interval, explaining 5.6% of the overall inequality. Factors such as maternal age at first marriage and maternal age at birth of the preceding child explained only a small percentage of the inequalities. Place of residence was the primary contributor to the reduction (-18.6%) of socioeconomic inequality in short birth interval followed by maternal education (-6.7%). Figure The value for the error term/residual component of the model was 0.0039. Since the residual approach to zero, the decomposition analysis provided a well-specified model where the socioeconomic-related inequality in short birth interval that was not explained by systematic variation in the determinants across wealth groups was too small .To our knowledge, this is the first study to quantify the socioeconomic inequality in short birth interval and its drivers in Ethiopia. The findings showed that short birth interval was more concentrated among the poor. Wealth quintiles, administrative regions, and not listening to the radio were the major contributors to the pro-poor socioeconomic inequalities in short birth interval. The findings of the current study can help policy makers and program planners address the inequality issue to achieve the SDG 10 target. Generally, this study is a valuable addition to the scientific literature on inequality and fills the knowledge gap on the socioeconomic disparities in short birth interval, which is one of the contemporary maternal health issues.In this current study, it was found that short birth interval is in favor of women from poor households. However, this does not imply that short birth interval does not occur among the rich. Rather it revealed short birth interval is disproportionately concentrated among the poor. Evidence has also shown that health and poverty are intimately intertwined, and poor health is more prevalent among the poor community than the better-off group , 59. TheIt was found that wealth quintiles itself (74.2%) were the predominant determinant of the overall socioeconomic inequality in short birth interval. Evidence from a cross-country study performed in 18 Sub-Saharan countries has also revealed that wealth was the single most important driver of inequality in maternal healthcare utilization . Prior sThe current study also demonstrated that administrative regions of residence were the second most important contributors (26.4%) to socioeconomic inequality in short birth interval. Literature has also shown that poverty is much more widespread and severe among pastoralists and the agro-pastorals region of Ethiopia due to their scattered and nomadic lifestyle , 67, whiWe found that place of residence contributed to the reduction (− 18.6%) of socioeconomic inequality in short birth interval. In spite of the government efforts to deploy a large number of community health extension workers and expand health facilities including health posts in most of the rural areas –69, the It was found that not listening to radio also contributed to the inequalities in short birth interval, explaining 5.6%. The 2016 EDHS has reported that radio was the most frequently accessed type of media . Other sAs a strength, the study used data from a nationally representative and large sample of a population-based survey. The application of a more rigorous decomposition analysis to determine factors contributing to socioeconomic inequality in short birth interval is another strength of the current study.We acknowledge the following limitations of this study. First, the cross-sectional nature of the study design does not allow us to draw causal inferences. Second, the use of wealth quintile as a measure of socioeconomic status should be considered cautiously. If the standardized living status measurement were used, some of the upper tail of the wealth distribution such as middle and richer wealth quintiles would fail below the poverty line.There was a pro-poor inequality of short birth interval in Ethiopia. Wealth quintiles, regions of residence, and not listening to the radio were the major contributors to the pro-poor socioeconomic inequalities in short birth interval.Households’ wealth status, administrative regions, and women’s media exposure particularly to the radio should be interventions priority targeting the reduction of socioeconomic-related inequality in short birth interval. Accelerating the implementation of poverty alleviation projects, such as implementing social safety net programs  and creaAdditional file 1: Supplementary Table 1. Key explanatory variables included in the decomposition analysis.