Patient adherence is defined as the extent to which a patient complies with medical or health advice. At present, there is a lack of reliable and valid measures specifically designed to measure adherence to internet-delivered behavioral interventions.The objective of this study was to develop and psychometrically evaluate a novel measure of adherence to guided internet-delivered behavioral interventions.In collaboration with experienced clinicians and researchers in the field, a 5-item, clinician-rated internet intervention Patient Adherence Scale (iiPAS) was developed. The initial scale was tested in a sample of children and adolescents (N=50) participating in internet-delivered cognitive behavioral therapy (ICBT) studies. A revised version of the iiPAS was then administered to a larger sample of children and adolescents (N=148) with various behavioral problems participating in ICBT trials. The scale was evaluated according to a classical test theory framework.The iiPAS demonstrated excellent internal consistency. Factor analyses revealed one underlying factor, explaining about 80% of the variance, suggesting that the scale captures a homogeneous adherence construct. The iiPAS was strongly associated with objective measures of patient activity in ICBT . Furthermore, mid- and posttreatment ratings of the iiPAS were significantly correlated with treatment outcomes. By contrast, objective measures of patient activity in the Web-based platform did not correlate with treatment outcomes.The iiPAS could be a useful tool to measure adherence in a broad range of internet-delivered behavioral interventions. Patient adherence, sometimes also termed compliance, is most broadly defined as “the extent to which the patient’s behavior  coincides with medical or health advice” . Within One important aspect of patient adherence is the logical assumption that adherent patients will have better treatment outcomes than nonadherent patients, though the evidence has not always supported this assumption. There is, for example, evidence that adherence to homework assignments in cognitive behavioral therapy (CBT) is associated with better outcomes, with a small to large effect of homework quality and quantity, but with substantial heterogeneity between studies ,9. SimilInternet-delivered behavioral interventions have emerged as an interdisciplinary development of psychotherapy and modern information technology. Internet-delivered behavioral interventions share common features as they mirror face-to-face psychotherapy regarding content and therapeutic techniques but are presented online via a personal login to a website or internet portal. The format usually resembles an online course, with instructive texts, pictures, videos, and written exercises, with the content structured similar to a book with chapters or modules. When therapist contact and feedback are involved, the treatment is referred to as guided or clinician-supported and otherwise as unguided or self-help . Interneshould experience the content (of the intervention) to derive maximum benefit from the intervention, as defined or implied by its creators” [Adherence may be particularly difficult to measure in internet-delivered behavioral interventions . In facereators” .Thus, defining and measuring the core therapeutic concepts and minimal effective use of those strategies may be a vital aspect of any measure of adherence to internet-delivered interventions. However, less than half of the studies in the field apply such a definition, and only few provide a justifiable definition of minimal required usage . CurrentIn summary, there is an important research gap and a need for reliable and valid measures of patient adherence to internet-delivered interventions. We take the perspective that a useful measure of adherence to internet-delivered behavioral interventions should (1) not only capture the structural aspects of the internet format (such as completed chapters or modules) but also consider the intended usage of its central therapeutic aspects  and (2) include relevant, observable behaviors that represent the patients’ adherence in a way that guides clinically meaningful decisions. Thus, the aim of this project was to construct and validate an adherence measure, evaluate its psychometric properties, and establish whether this new measure correlates better with treatment outcome than other traditional measures of adherence .A novel, clinician-rated measure called internet intervention Patient Adherence Scale (iiPAS) was constructed in a 4-step process following recommendations from the psychometric literature ..The first author (FL) conceptualized the scale and drafted a first version with the aim to measure patient adherence as one homogeneous construct. To ensure feasibility, usability, and precision, the scale was intended to be a short and time-efficient measure and assess patient behaviors that would be observable by clinicians. The context of the scale was defined as a clinician-rated measure to be used within clinician-supported, internet-delivered behavioral interventions. The advantage of a clinician-rated scale, compared with a self-rated measure, was justified by the ability of the clinician to assess the adherence of the patient relative to the intended, optimal use of the intervention, which was assumed crucial for the reliability of the instrument . BuildinTo ensure optimal content validity, experienced clinicians and researchers with relevant competence (see “Expert Sample”) were involved in choosing the response format and initial item formulation. The initial draft was circulated among the experts in 2 review rounds and edited iteratively by the first author according to suggestions and comments. The preliminary version of the iiPAS included 5 clinician-rated items rated on a 3-point Likert scale, including anchor labels for the end points of the scale (see description under “Measures”).The preliminary iiPAS was tested in a small sample (“Patient Sample 1”) to provide basic psychometric information as well as to give clinicians the possibility to test the scale and give feedback regarding usability and applicability.On the basis of additional feedback from step 3, the wording of the items was minimally revised for improved clarity. In addition, the 3-point Likert scale was extended to a 5-point scale to increase response variance. A preliminary psychometric analysis was carried out; results are presented below. For the full psychometric evaluation, the revised iiPAS was then administered to participants in several clinical trials of internet-delivered behavioral interventions (“Patient Sample 2”).The expert sample consisted of 8 experienced researchers and clinicians who had been involved in clinical trials of internet-delivered behavioral interventions  within a Swedish research network called the Child Internet Project or Barninternetprojektet (BiP). The experts were involved in the initial scale development.Diagnostic and Statistical Manual of Mental Disorders, 4th edition (DSM-IV) and a clinical severity score above 3 on the Clinician Severity Rating (CSR) [Initial data were collected from 50 patients, of whom 33 (66%) were children participating in a study of ICBT for pediatric anxiety , and 17 ng (CSR)  or a pring (CSR) . Clinicing (CSR) ,25. The Patient sample 2 consisted of 148 children and adolescents with various mental disorders, who together with their primary caretakers participated in different clinical trials. Common inclusion criteria within all trials were (1) age within the age interval for the respective trial (see “Interventions”), (2) internet access, (3) absence of suicidality or another severe condition that required immediate attention or higher priority, and (4) the clinical presentation of a mental disorder that was targeted in the specific trial . All conditions were diagnosed by experienced psychologists using semistructured diagnostic interviews based on DSM-5 criteria . The iiPAll interventions were provided by researchers and clinicians associated to the BiP network, a collaboration between the Karolinska Institutet and the Child and Adolescent Mental Health Service in Stockholm, with a main research focus on development and evaluation of internet-delivered interventions for childhood mental disorders. All interventions in the current evaluation were clinician-guided ICBT interventions that also included parent support. Clinician advice, support, and feedback were given predominantly via written messages within the secure internet platform or via occasional telephone calls. For a short summary of the interventions, see The included clinical trials applied similar but not identical procedures. A description of the common aspects of the procedures is provided here; for a detailed description, see the respective trial publications. Participants in patient samples 1 and 2 were recruited via clinician- or self-referrals. Selection of eligible participants included both telephone screenings with a clinician and an in-person assessment with the child or adolescent and at least one caregiver. Fulfillment of diagnostic criteria was established by a clinician  using the Mini-International Neuropsychiatric Interview child version (MINI Kid ) or the The BiP TIC and BiP Anxiety trials had a randomized controlled design ,29. In BThe clinical outcome measures were assessed at pre- and posttreatment. Activity data from the use of the internet platform  were automatically collected during the treatments. The iiPAS was administered by the clinicians participating in the trials, on 2 separate occasions: halfway through the treatment (iiPAS-mid) and at posttreatment (iiPAS-post).The iiPAS is a 5-item, clinician-rated measure of patient adherence to internet-delivered behavioral interventions. The 5 items are rated by treating clinicians and cover 5 central aspects of adherence: the patient’s work pace, engagement, communication with the clinician, motivation for change, and login frequency. Each item is rated on a 0 to 4 Likert scale. Thus, the iiPAS-total score ranges from 0 to 20, with 0 indicating no adherence and 20 perfect adherence. The items are designed to generically suit different types of internet-delivered behavioral interventions. A prerequisite, however, is that the intervention is guided, that is, clinician-supported, as opposed to unguided or entirely self-help. The scale is constructed to be applicable to children and adolescents as well as adults. In the case where caregivers are involved in the patient’s treatment, the iiPAS is primarily intended to evaluate the patient’s adherence to treatment rather than the parent’s or carer’s involvement. See Clinical Global Impression-Severity (CGI-S ) is a co), 8 maximal severity, and 4 the cut-off for clinical severity . ADIS has previously shown excellent interrater reliability [CSR is rated in conjunction with the clinician-administered ADIS , which iiability  and testiability .The internet treatment platform that the patients logged in to for accessing treatment material  provided user data on number of logins to the platform, duration of logged in time, and number of completed chapters and written characters by the patients in communication and exercises. The number of completed chapters was expressed in percentage relative to total number of chapters because of the different number of chapters in different interventions.Cronbach alpha was calculated as a measure of internal consistency. An exploratory factor analysis (EFA) was conducted regarding the factor structure of the iiPAS, including the Kaiser-Meyer-Olkin (KMO) measure of sampling adequacy to determine the adequacy of the data for factor analysis as well as a Bartlett Test of Sphericity to determine the factorability of the five items.Pearson correlations were used to explore convergent validity, defined as the association of the iiPAS with platform usage variables. In variables where outliers with large leverage were identified, Spearman correlation coefficient was used. The intercorrelations of iiPAS between the halfway treatment time point (iiPAS-mid) and at the end of treatment (iiPAS-post) were calculated. For some calculations, an iiPAS total score was used, which is the sum of iiPAS-mid and iiPAS-post, to obtain a measure of adherence that represents the whole treatment period from start to end of treatment. Regarding criterion validity, the association between iiPAS and symptom severity change from pre- to posttreatment was calculated. Because the available measures, CSR (in the BiP Anxiety group) and CGI-S (in the remaining groups), are similar clinician-rated, 1-item Likert scales, but not identical, the CSR and CGI-S change scores were z-transformed and then combined into a single new variable (“symptom severity change”), which was then further used in the analyses. There were only a few missing values in the dataset (2.7%), and missingness was, therefore, not further handled but listwise excluded from the respective analyses. Analyses were conducted in SPSS Statistics version 24 (IBM Corp) and R (R Core Group ).An initial version of the iiPAS was administered to 50 children and adolescents, aged between 7 and 17 years, receiving ICBT for OCD or anxiety disorders. Preliminary analyses of Cronbach alpha indicated excellent internal consistency .The majority of the 148 participants were female , with a mean age of 12.7 years. 210=578.5, P<.001 indicated factorability. The results of the EFA indicated a single factor with an eigenvalue of 3.97 that explained 79.3% of the variance. For iiPAS-post, KMO was 0.85 and Bartlett Test of Sphericity χ210=787.0, P=.001, again indicating factorability. The EFA resulted in a single factor with an eigenvalue of 4.18 that explained 83.6% of the variance. For iiPAS-mid, a KMO=0.89 and Bartlett Test of Sphericity χr=0.25-0.79), apart from nonsignificant correlations between iiPAS and logged in time. There was also a substantial and significant correlation between iiPAS-mid and iiPAS-post (r=0.78). Visual inspection of the scatterplots did not indicate any nonlinear associations. iiPAS-mid, iiPAS-post, and iiPAS-total scores were positively and significantly correlated with platform usage variables , and item 4 .iiPAS-mid, iiPAS-post, and iiPAS-total scores correlated positively and significantly with the standardized symptom severity change score  is less important for the clinical outcome than actively engaging with the therapeutic content that is presented in the intervention . The existing literature on patient adherence supports that quality of homework exercise completion is a better predictor of treatment outcome than number of completed exercises ,13. HoweOur analyses included diverse conditions, ranging from anxiety disorders, OCD, tics/Tourette’s disorder, and self-harming behaviors. Future studies should aim for an even broader evaluation of the iiPAS in additional diagnoses and symptoms that are currently treated with internet interventions, especially in the behavioral medicine field, where about half of the internet treatment trials are conducted . MoreoveImportantly, the iiPAS was designed to be useful in a wide range of internet-delivered, clinician-supported internet interventions of different therapeutic schools. However, the interventions in this study are all based on a CBT framework. An important next step would be to test the iiPAS in different internet-delivered behavioral interventions. Regarding the reliability analyses, an important aspect that was omitted in this study was the calculation of interrater reliability. Due to the scoring format of the iiPAS, this would require 2 clinicians for each patient, which was not feasible at the current stage of evaluation but would be important to take into account in future studies.Finally, the sample size in this study was big enough for a global evaluation but too small to conduct detailed separate analyses for each specific condition or age group. A larger sample would allow analysis of the association between patient adherence and clinical outcomes in greater detail with regards to specific patient characteristics such as age, comorbidity, and symptom severity at baseline.To summarize, the iiPAS demonstrated sound psychometric characteristics in a clinical sample of children and adolescents with a broad range of psychiatric and behavioral problems. The iiPAS appears to be a valuable tool to measure and study different aspects of adherence within the context of clinician-guided, internet-delivered behavioral interventions. Further evaluations within different age groups, types of internet-delivered interventions, and patient populations are warranted.


Chromosomal disorders, of which Down syndrome is the most common, can cause multi-domain disability. In addition, compared to the general population, there is a higher frequency of death before the age of five. In many settings, large gaps in data availability have hampered policy-making, programme priorities and resource allocation for these important conditions. We have developed methods, which overcome this lack of data and allow estimation of the burden of affected pregnancies and their outcomes in different settings worldwide. For example, the methods include a simple equation relating the percentage of mothers 35 and over to Down syndrome birth prevalence. The results obtained provide a starting point for consideration of services that can be implemented for the care and prevention of these disorders.The online version of this article (10.1007/s12687-017-0336-2) contains supplementary material, which is available to authorized users. Chromosomal disorders are caused by changes occurring in either chromosome number or structure usually during the formation of gametes or soon after fertilisation. These changes can affect the autosomes or the sex chromosomes , with chromosomal disorders divided into the two corresponding groups. Most autosomal disorders cause death before the age of 5 years or multi-domain disability. However, the availability of appropriate medical care in high-income settings means that there are fewer deaths in those with Down syndrome before the age of five  uses a set of defined methods to relate demographic data to the known birth prevalence of selected congenital disorders, to generate estimates relevant to public health, policy making and clinical practice ) in the no-care situation and with current care. The starting point for this is baseline prevalence (i.e. live birth and stillbirth) in the absence of care. MGDb provides epidemiological estimates relating to birth prevalence, pregnancy outcomes, survival and effect of interventions. These estimates are not definitive and require refinement, but can be used by those working in public health or by policy makers as a starting point to assess service needs and policy gaps in settings with little or no data.Step 1: Estimation of the expected prevalence of Down syndrome among live births in the absence of any intervention using the established relationship with maternal ageStep 2: Estimation of the birth prevalence of other chromosomal disorders and foetal deathsStep 3: Estimation of the overall baseline birth prevalence of chromosomal disorders in the absence of any interventionStep 4: Estimation of actual birth prevalence based on the availability and estimated access to prenatal diagnosis and termination of pregnancy (TOP) for foetal impairmentStep 5: Comparisons of estimates with registry dataTo estimate the potential live birth prevalence of chromosomal disorders for different populations, we used a step-wise approach:The live birth prevalence of chromosomal disorders in the absence of intervention was originally described in populations mainly of northern European origin xpercentage of mothers ≥ 35 × 0.067bthe intercept (0.834)If age heaping is not taken into account, it can lead to inaccuracies in the estimation process and overestimation of the live birth prevalence of Down syndrome. We sought to reduce the effect of heaping by developing a formula to describe the relationship between the percentage of mothers aged ≥ 35 and Down syndrome live births. This was undertaken in two steps. Firstly, we used the rates for Down syndrome birth prevalence in Table A comparison of estimated potential live birth prevalence, applying the risks to the five-year bands, compared to the formula above using WPP 2015 maternal age data from 1950 to 2015, can be found in the online resources for three countries  used in MGDb was derived from an average of country-specific rates using publicly accessible EUROCAT data for the years 2000–2009. Data included in the analysis was from registries that submit raw data and from countries where TOP is legal and reported . Average EUROCAT data indicate that the potential live birth prevalence of Turner syndrome  is approximately 0.175/1000.The birth prevalence of Klinefelter syndrome (XXY) is weakly related to maternal age because around half of the responsible non-disjunctions are of paternal origin  are associated with chromosomal aneuploidy and risk of miscarriage accordingly rises with maternal age  have recorded almost the entire evolution of the population impact of screening and diagnosis for chromosomal disorders in high-income settings. Consequently, country-specific data on termination of pregnancy from umbrella registries are used for these estimates when available. Otherwise, we have used publicly available data from EUROCAT to estimate the average proportions of pregnancies terminated by diagnosis by 5-year interval, in participating countries where termination of pregnancy is legal and reported Table . InformaWe compared these rates with those of the UK National Down Syndrome Cytogenetic Register  ) in ordeDue to the absence of readily accessible observational data on the availability and utilisation of prenatal diagnosis, we have assumed that it will be available in those countries where TOP is legal and access to care will impact utilisation. In MGDb, we have used the EUROCAT average rates for settings that have a policy of universal screening. In settings where termination of pregnancy is legal but there is no evidence of a universal screening policy and no available data on termination rates, we have assumed testing is offered only to high-risk groups. In these settings, we have used a rate which is 50% of the EUROCAT average. We have assumed that all women worldwide with access to screening and TOP will act in the same way in terms of uptake of this service. It is assumed that no widespread screening for chromosomal disorders or terminations takes place in countries where termination of pregnancy for foetal impairment is illegal. This is likely to lead to slight overestimation of actual live birth prevalence, as a minority of couples can usually access such services, for example as in the case of Brazil  would not greatly affect the reported birth prevalence of Down syndrome. Under-ascertainment seems more likely for terminations than for live births, particularly where there is widespread use of private maternity services or termination of pregnancy is illegal or a highly politicised issue. Under-ascertainment of terminations for other trisomies has been documented in the USA  as well as at the proportion of the total population they cover. This can lead to differences between countries in reported rates, especially for outcomes that do not end in an affected birth such as foetal death/stillbirth and termination of pregnancy. Secondly, there is a high rate of TOP for Down syndrome, other autosomal disorders and Turner syndrome in high-income countries where termination is legal. This assumption may therefore overestimate TOPs in other settings and therefore underestimate the number of foetal deaths and live births. Termination may be under-reported from countries where termination for foetal impairment is legal, but the issue is highly politicised. The extent to which termination reflects, informed parental choice has not been sufficiently investigated. Estimation of termination of pregnancy is based on EUROCAT and ICBDSR registry data. However, there may be considerable under-ascertainment in some countries


Randomization was computer-generated by the statistic program Epidat v.4.1 and allocation was performed using sealed opaque envelopes. Clinical measurements and peripheral blood samples for biochemical variables were collected at baseline, in the middle of second trimester before non-surgical treatment, and in the third trimester. Microbiological samples were collected in the second and third trimester. A statistically significant reduction was verified in all clinical and microbiological parameters after periodontal treatment in the test group. No significant differences were observed for the rest of the variables, including preterm birth and/or low birth weight. No adverse events related to periodontal treatment were reported. Conclusions: Non-surgical periodontal treatment in Caucasian patients with periodontitis stage II grade B did not significantly reduce the risk of adverse pregnancy outcomes.Aim: To analyze if non-surgical treatment of periodontitis in a pregnant Caucasian women population can reduce adverse pregnancy outcomes. Methods and results: A parallel randomized clinical trial was designed and approved by the Ethical Committee of Sanitary Area Santiago-Lugo, Spain (registration number: 2016/451). Forty patients with periodontitis stage II grade B were randomly allocated to receive either comprehensive non-surgical periodontal therapy (test group;  Adverse results of pregnancy, such as preterm birth or low birth weight, can be the cause of mortality and increased morbidity in neonates . WorldwiThe prevalence of preterm births varies depending on the country (ranged from 5% in some European countries to 18% in some African countries). Blencowe et al.  identifiFusobacterium nucleatum, Treponema denticola, Tannerella forsythia, Campylobacter rectus, Eikenella corrodens, and Selenomonas sputigena [Periodontitis is an inflammatory infection disease affecting the tooth-supporting tissues, initiated by a biofilm with gram-negative anaerobic microorganism predominance and mediated by the inflammatory response of the host. It is also liable to induce systemic inflammation and consequently modify the normal development of pregnancy . During putigena .Fusobacterium nucleatum, Campylobacter rectus, Porphyromonas gingivalis, and Bergeyella spp. [Infections account for 75% of premature births and/or low birth weight, while subclinical infections begin to be considered as an important cause of very premature birth (before the 30th week of gestation). However, in more than 50% of cases, the etiological factors of prematurity are unknown . Oral inlla spp. .Two main routes have been proposed to trigger an inflammatory response in the fetoplacental unit; the direct route, in which the oral microorganisms and/or their components reach the fetoplacental unit by hematogenous dissemination of the oral cavity or by an ascending route through the genitourinary tract; and the indirect route, in which the inflammatory mediators produced in the periodontal tissues circulate and impact the fetoplacental unit. A large number of studies associate an increase in the levels of local and systemic inflammatory markers with adverse pregnancy outcomes . As wellThe increase in maternal serum levels of proinflammatory cytokines, such as IL-1, IL-6, IL-8, and TNF-α have been associated with prematurity and low birth weight ,14. C-reCohort studies have shown a relationship between periodontal disease, premature delivery, and low birth weight ,17,18, aOur hypothesis was that periodontitis has an effect on pregnancy outcomes, such as preterm birth and/or low birth weight. Therefore, the aim of the current study is to analyze if non-surgical treatment of periodontitis, using the definition suggested by the EFP-AAP Consensus Meeting , can redA parallel randomized clinical trial was designed to test whether the non-surgical periodontal treatment provided during pregnancy affected gestational age at delivery and birth weight. Data were gathered on the pregnant women at three visits: at the end of the first trimester (16 weeks of pregnancy), in the middle of second trimester (22–24 weeks of pregnancy), and at the third trimester (33–36 weeks of pregnancy), as usually recommended follow-up visits in pregnant woman. No relevant changes were performed after trial commencement.Pregnant women who were seen for prenatal care at the University Hospital of Santiago de Compostela (Spain) were invited to participate in the study when they attended the hospital to undergo the first gynecological clinic appointment between June 2017 and May 2018. During this screening visit, a full-mouth periodontal examination was performed and patients with a periodontitis diagnosis  were invInclusion criteria were Caucasian pregnant women , aged 18 to 40 years, with more than 20 natural teeth and diagnosis of periodontitis. Exclusion criteria were multiple gestation; previous preterm low birth weight; more than 1 previous miscarriage or one greater than 18 weeks; diabetes; hypertension; alcoholism; drug abuse; human immunodeficiency virus (HIV) infection; heart disease, kidney disease or liver disease; recurrent cystitis; viral infections; venereal infections; toxoplasmosis; and/or tobacco smoking.Demographic data and medical history were assessed by interview during the first visit . The interview included items on age, educational level , area of residence , body mass index (BMI), marital status , and obstetric history . The following variables were recorded shortly after delivery: newborn weight, newborn sex, duration of pregnancy, and type of delivery . Interdental CAL was detectable at ≥2 non-adjacent teeth, orBuccal or oral CAL ≥3 mm with pocketing ≥3 mm was detectable at ≥2 teeth but the observed CAL could not be ascribed to non-periodontitis-related causes such as: (1) gingival recession of traumatic origin; (2) dental caries extending in the cervical area of the tooth; (3) presence of CAL on the distal aspect of a second molar and associated with malposition or extraction of a third molar; (4) an endodontic lesion draining through the marginal periodontium; and (5) the occurrence of a vertical root fracture.Full-mouth periodontal examinations were performed by one examiner. The weighted κ values for intra-examiner calibration were 0.82 (CI 95% = 0.68–0.98). Periodontal clinical measurements were recorded at enrolment and repeated at 2nd and 3rd trimester. Exams included all teeth present in the mouth (excluding third molars). A plaque score , periodoA peripheral blood sample was collected from each subject with venipuncture using a vacuum tube . Samples were centrifuged at 2500 rpm for 10 minutes and the serum obtained was dissociated in a plastic tube. At the time of the analyses, the serum concentration of interleukin-6 (IL-6), interleukin-8 (IL-8), tumor necrosis factor-α (TNF), and fibrinogen were determined using commercially available enzyme-linked immunoassays (ELISA). In the middle of the second trimester (22–24 weeks of pregnancy) before non-surgical periodontal treatment and before the third trimester, four sites (those with the deepest probing depth) were selected, and two consecutive paper points were inserted and kept in place for 10 s. All paper points were pooled in a vial with reduced transport fluid (RTF) and transferred to the laboratory within 12 h .2; 10% H2; 10% CO2) at 37 °C, and after 3–5 days at 37 °C in air with 5% CO2.The obtained samples were dispersed (30 s of vortex), serially diluted and processed for culture by inoculation on two different media: blood agar medium , with horse serum at 5% and with hemin (5mg/L) and Dentaid-1 medium . The bloPorphyromonas gingivalis, Prevotella intermedia, Tannerella forsythia, Parvimona micra, Campylobacter restus, Fusobacterium nucleatum, Eikenella corrodens, and Capnocytophaga spp., were identified based mainly on their colony morphology and further confirmed with different specific chemical tests. Counts for every bacterial species were obtained and the percentage relative to the total flora calculated. Aggregatibacter actinomycetemcomitans were grown on Dentaid-1 medium plates, and identified based on colony morphology and catalase reaction.The test group patients received non-surgical periodontal therapy that was completed by the end of week 24 of gestation. This consisted of oral hygiene instructions, followed by mechanical supra and subgingival scaling and root planning (SRP) with Gracey curettes . Periodontal therapy was performed over two 1-hour sessions. Periodontal treatment was performed by one periodontist (L.C.) in the Department of Periodontology at the University of Santiago de Compostela (Spain).The control group received professional tooth cleaning  at the second-trimester visit. All control participants were offered the opportunity to attend for non-surgical periodontal therapy postpartum.Primary outcomes of the present study were preterm birth and low birth weight. Following the World Health Organization criteria, preterm birth was defined as a delivery at <37 weeks of gestation  and low birth weight was defined as a newborn weight of 2500 gr or less. Secondary outcomes were those related to the efficacy of non-surgical periodontal therapy on clinical, biochemical and microbiological variables.p-values <0.05 for all analyses were selected to be statistically significant. SPSS for Windows  was used for all the statistical analyses. Descriptive analysis was calculated for each variable . Kolmogorov–Smirnov test was performed to test the normality of the variables. Relationship between each variable and preterm birth/low birth weight were analyzed. All variables were also compared between women with and without treatment. A t-student test was performed for continuous variables whilst A chi-square test was used to analyze categorical variables. Non-normal distribution variables were analyzed by a Mann–Whitney U test for continuous variables and chi-square test for categorical variables. Multivariate analysis was performed using ANCOVA test, with pregnancy and newborn weight as dependent variables with age, body mass index as a covariable, and treatment as an independent variable. Statistical significance was established at the 95% confidence level and Initially, 66 women were examined and 50 met the inclusion criteria. Eight rejected to participate and two had a miscarriage. Finally, 40 women with periodontitis stage II grade B complied with all the visits, 20 received non-surgical treatment (test group), and 20 received no treatment was performed (control group), as shown in Demographic characteristics of the study sample are shown in Porphyromonas gingivalis, Fusobacterium nucleatum, Prevotella intermedia, and Tannerella forsythia. No statistically significant differences were observed between the control and test group. The microbiological findings from 40 pregnant women, showed high proportions of detection of The mean gestational duration was 38.25 ± 2.88 weeks for the control group and 37.72 ± 6.04 weeks for the test group. Mean weight at birth were 3012.59 ± 415.42 g for the control and 3249.48 ± 473.54 g for the test group. No statistically significant differences were observed between groups for these variables.The main objective of this study was to evaluate the effects of non-surgical periodontal treatment in pregnant women, carried out during the second trimester of pregnancy, on the rates of preterm birth and low birth weight, and to assess the clinical, biochemical, and microbiological changes. The effect of periodontal intervention on pregnancy outcomes did not show a reduction in the risk of preterm birth and low birth weight in the test group compared to the control group. However, we can assume as clinically relevant the protector role of periodontitis treatment during pregnancy , due to the significant periodontal health improvement in the test group. Pregnancy is a short period of time in order to evaluate the systemic impact of a long term periodontal microbial infection and related inflammatory responses. This finding is supported by previous studies such as Oliveira et al. , OffenbaConversely, some papers have suggested a positive effect of non-surgical periodontal treatment both in the periodontal condition of the patient and in the adverse results of pregnancy. Offenbacher et al.  showed tWhen the clinical variables were analyzed, we have to consider that bleeding on probing, probing pocket depth, and clinical attachment level were statistically higher for the test than for the control group in the 1st trimester, however both groups were diagnosed with periodontitis stage II, grade B . These cBiochemical markers increased in the third trimester following the basic periodontal treatment, except TNF-α, which decreased after performing the treatment. The differences between the treatment group and the control group were not significant for any of the three visits. Most authors agree that treatment of chronic periodontitis does not significantly change levels of serum markers of acute-phase inflammatory and vascular response . It has Fusobacterium nucleatum, Prevotella intermedia, Porphyromonas gingivalis, and Parvimona micra [Eikenella corrodens with preterm birth and the presence of Capnocytophaga spp. with low birth weight. Novak et al. [Regarding the presence of periodontopathogens, the non-surgical periodontal treatment performed in the second trimester reduced the number of all the subgingival periodontal pathogens analyzed and the total bacterial load. These findings indicate that the periodontal therapy of pregnant women with periodontitis produces quantitative and qualitative changes in the microbiota similar to those observed in non-pregnant women . Pregnanna micra . Martínena micra  did not na micra  found a k et al.  showed tThe present study is not exempt from a number of methodological limitations that should be considered when extrapolating our results. Up to date, available evidence does not provide clinical, microbiological, or immunological parameters that define patients whose pregnancy outcomes would be improved with periodontal treatment . The selWe can conclude that non-surgical periodontal treatment in Caucasian patients with periodontitis stage II grade B did not significantly reduce the risk of adverse pregnancy outcomes.


The crude Odds Ratios (OR), adjusted (aOR) and their 95% confidence intervals (CI) were calculated through binary logistic regression. For multiparous women in private centers, a higher rate of induced labor was observed , fewer natural methods were used to relieve pain , and increased odds of cesarean section  were found as compared to public hospitals. For primiparous women in private centers, a greater use of the epidural was observed , as well as an increased likelihood of instrumental birth  and of cesarean section  than in public hospitals. No differences were found in hospitalization times among women giving birth in public and private centers (p > 0.05). The World Health Organization birth attendance recommendations are more strictly followed in public hospitals than in private settings.The type of hospital (public or private) has been associated with the type of clinical practice carried out. The purpose of this study was to determine the association between the type of hospital (public or private) and delivery attendance with practices based on the recommendations by the World Health Organization (WHO). A cross-sectional study with puerperal women ( The World Health Organization (WHO), in the Fortaleza Declaration, provides birth attendance recommendations to promote non-medicalized assistance without interventions . To a grThe main care supplier regarding birth attendance in most health systems worldwide is the public health administration ,11,12,13To date, different studies have associated the type of center with the skilled birth attendance recommendations by the WHO, based on the private or public character of centers ,22,23,24Despite three decades having elapsed since the WHO made its birth attendance recommendations, which are mainly aimed at reducing the rate of cesareans, among other objectives, existing studies reveal the reluctance of private hospitals to include them in their clinical practice. In recent years, a twofold trend has been observed: on the one hand, scientific evidence supports more humanized birth attendance while, on the other hand, more women choose a private hospital to give birth. The aim of this study was to identify whether there is an association between the type of hospital where the woman gives birth (public or private) and the performance of the clinical practices recommended by the WHO on attendance at delivery.A cross-sectional study was conducted with a total sample of 3437 women who gave birth in Spain in 2017.The inclusion criterion was singleton term pregnancies. The exclusion criteria included pregnancies that ended in antenatal fetal death, giving birth at home, breech delivery, and those women whose birth was initially attended to at home and ended in a hospital.We estimated the appropriate sample size according to the results of a similar study . The likAs this is an observational study, in order to know the prevalence of birth attendance in both types of centers, the recruitment of women for the study was not restricted to a given number. Instead, the recruitment time was limited to three months, that is, the data collection was carried out during three months, and all the women who gave birth in that period of time and who met the inclusion criteria and wished to participate were included. An online questionnaire devised by the authors was used for data collection. It comprised 35 items (three open and 32 closed questions) about socio-demographic and clinical characteristics, obstetric results, and data on newborns. The questionnaire was previously trialed and distributed to women via the main women associations, the Spanish Federation of Midwives Associations (FAME) and its member associations that engaged midwives in the dissemination of the project and encourage women to participate. After the study subjects’ selection and upon their acceptance to participate, the midwives in charge of recruiting the women provided them with instructions to fill in the questionnaire, which the final participants did according to their availability. Telephone assistance and a scheduled chat were provided to clarify any doubts these women may have when filling the questionnaire.The main independent variable was the type of center where birth attendance was provided (private/public). The main outcome variables were: induced birth, preparing a birth plan, employing local analgesia , employing natural methods to relieve pain, application of fundal pressure during the second stage of labor (Kristeller maneuver), cesarean, re-hospitalization after hospital discharge, postnatal surgery, mother admitted to an intensive care unit (ICU), skin-to-skin contact at birth, newborn admitted to hospital, feeding/milk type, and length of hospital stay. The following were also selectively evaluated as outcomes for women who gave birth vaginally: birth type , practicing episiotomy, and presence of severe tearing (grade III or IV).The secondary variables taken into account to control confounding factors were of a socio-demographic and clinical type. In addition, the variables that could potentially act as confounders were used for each outcome.U test was used. A p ˂ 0.05 was considered significant. Additionally, all the analyses were done using the SPSS v24.0  statistical package.First, a descriptive analysis was carried out for which absolute and relative frequencies were used. Then, a bivariate analysis between the type of center and the main obstetric and neonatal results was done by stratifying by parity (primiparous/multiparous): the odds ratios (OR) were calculated, along with the 95% confidence interval (95% CI) and the Pearson’s chi-squared test. Then, a multivariate analysis was performed by means of binary logistic regression, which was stratified by parity. In addition, the potential confounding variables were used for each analysis. Finally, both types of center were compared regarding the length of hospital stay by stratifying by birth type, for which a non-parametric Mann-Whitney This study was approved by the Ethical Committee on Clinical Research  with ethical code 69-C of the La Mancha-Centro Centre. Before starting the questionnaire, the participating women read a fact sheet about the study, its objectives, etc., and marked a box by which they showed their consent to participate in it, i.e., they signed an online informed consent (ticking the option if they wanted to participate or not doing so when refusing to take part in the study).p < 0.001). In addition, multiparous women giving birth in public centers were associated with attendance to antenatal childbirth classes (p < 0.001) and the presence of health problems during pregnancy (p = 0.043).2906 women were recruited, of whom 596 attended private hospitals and 2310 public ones. Therefore, public hospitals received 79.5% of the study women and 20.5% of them attended private ones. Those primiparous women who gave birth in private hospitals were also associated with a more frequent use of epidural  than their counterparts in public hospitals. Moreover, primiparous women who gave birth in private hospitals were associated with a greater likelihood of having an instrumental birth  or cesarean section , as opposed to those who gave birth in public hospitals.Regarding multiparous women who attended private centers, we observed a significantly higher rate of induced births , a greater use of analgesia , fewer natural methods to ease pain , and more births that ended in cesarean  .p = 0.226), instrumental (p = 0.988), or ending in a cesarean section (p = 0.101).As According to the results found in this study, birth attendance in private hospitals entailed more cesareans for both primiparous and multiparous women, and with a greater use of epidural analgesia. More primiparous women had instrumental births in private hospitals, but fewer women underwent episiotomies. In addition, skin-to-skin contact with their babies took place earlier than in public hospitals. More multiparous women births in private centers started in an unspontaneous way, and the professional staff resorted to fewer natural methods to relieve pain than in public centers. No differences were found between public and private hospitals regarding the length of post-birth hospital stay as for any birth type.It was decided to exclude premature newborns in order to achieve standardized comparisons between the centers. The incidence of premature newborns is highly variable depending on the level of care of the centers. Several limitations are identified in this study. A priori, it is worth thinking that if a selection bias exists and is associated with not participating in the study, this has not influenced the results, as the majority of women agreed to participate and only 30 refused doing so. Nor was there any reason to believe that those who did not participate would have had performed differently to those who did. It is unlikely that an information bias exists as the collected data and the way the possible answers were presented did not require having a high level of education. In this respect, the questions were set out in a basic and simple way, and could be understood by all the participants regardless of their level of education. On the other hand, it is not possible to completely rule out an anamnesis bias. Despite the information being collected over a short time interval, we believe that this would have minimally influenced the results. Moreover, women perfectly knew their health supplier and were able to indicate if they had been assisted in a private or public hospital. They also accurately remembered any information regarding their birth process, which was accordingly evaluated by the vast majority and regarded as a relevant process to devote attention to. Likewise, it is not possible to completely rule out the confounding bias inherent to observational studies. However, we believe that its effect on the study results was not significant as it was considered and controlled during the study design. This was done by using inclusion and exclusion criteria, as well as a multivariate analysis adjusted by confounding factors  that could have influenced the results. Another limitation is that the data used in the study were of an individual nature for each participant, so it cannot be assured that these outcomes act as indicators and results per center. In other words, it was not possible to find private centers with better results than public ones. However, the present study analyzed them overall.More women who gave birth in private centers had a cesarean section than those who gave birth in public hospitals, regardless of being primiparous or multiparous. This finding agrees with other authors’ results ,19,20,22On the one hand, our results also revealed that multiparous women who decided to give birth in private hospitals were more likely to have an induced birth, which contrasts with the results obtained by Wilkes et al. . On the During the study process, a greater likelihood of instrumental birth was found for primiparous women in private centers, along with regional analgesia being more frequently used . These results coincide with other authors’ outcomes ,19,30,31Primiparous women giving birth in private hospitals came over as a protective factor against performing episiotomy, unlike the results obtained by Escuriet et al. in their study conducted in Spain  and thosEarlier skin-to-skin contact was also identified in women giving birth in private hospitals than in those who did it in public ones, but no influence was detected on any other studied practice or intervention: moment of starting breastfeeding, newborn being admitted to hospital, presence of severe perineal tearing, mother being admitted to an ICU, length of hospital stay, mother being readmitted to hospital, performing Kristeller maneuver during expulsive contractions, performing surgery, and women using a birth plan. Some of the analyzed practices, e.g., using a birth plan, have been studied by Prado et al. , with rePrivate centers policies should include the recommendations by the WHO on birth attendance as a mandatory standard to be followed by the professionals working in these hospitals. In addition, the staff should be trained in and sensitized on the importance of carrying out childbirth attendance based on the WHO recommendations.To conclude, clinical practices regarding birth attendance in public hospitals do follow the WHO recommendations and the available scientific evidence to a greater extent than private hospitals. Thus, giving birth in a private center is associated with an increased likelihood of delivery by cesarean and use of epidural analgesia during delivery. In addition, primiparous women have been associated with a greater number of instrumental deliveries and less early skin-to-skin contact with the newborn. Finally, multiparous women show a higher incidence of induced labor.


Endocrine disrupting chemicals (EDCs) are ubiquitous, and pregnancy is a sensitive window for toxicant exposure. EDCs may disrupt the maternal immune system, which may lead to poor pregnancy outcomes. Most studies investigate single EDCs, even though “real life” exposures do not occur in isolation. We tested the hypothesis that uniquely weighted mixtures of early pregnancy exposures are associated with distinct changes in the maternal and neonatal inflammasome. First trimester urine samples were tested for 12 phthalates, 12 phenols, and 17 metals in 56 women. Twelve cytokines were measured in first trimester and term maternal plasma, and in cord blood after delivery. Spearman correlations and linear regression were used to relate individual exposures with inflammatory cytokines. Linear regression was used to relate cytokine levels with gestational age and birth weight. Principal component analysis was used to assess the effect of weighted EDC mixtures on maternal and neonatal inflammation. Our results demonstrated that maternal and cord blood cytokines were differentially associated with (1) individual EDCs and (2) EDC mixtures. Several individual cytokines were positively associated with gestational age and birth weight. These observed associations between EDC mixtures and the pregnancy inflammasome may have clinical and public health implications for women of childbearing age. Data from the National Health and Nutrition Examination Survey (NHANES) show that the vast majority of US adults have evidence of exposure to phthalates and phenols, such as bisphenol A (BPA) and polyfluoroalkyl chemicals, as measured in urine samples6. Children also demonstrate exposure to a wide variety of environmental chemicals8. EDC exposure may occur on a daily basis via packaged foods, plastics, cosmetics, and pharmaceuticals, and industrial applications may lead to widespread environmental exposure to heavy metals10.Environmental toxicants, which include endocrine disrupting chemicals (EDC) such as phenols, phthalates, metals, and organochlorines, are globally ubiquitous and represent an area of major public health concern13. Epidemiologic studies have demonstrated that EDC exposures are nearly universal in pregnant women18. Exposure to EDC classes including phenols, phthalates, parabens, flame retardants, and heavy metals may occur in pregnancy by way of personal hygiene products, cosmetics, household cleaning products, the use of electronic devices, and consumption of animal, plant, or processed foods20.EDC exposures are of particular significance to pregnant women, as fetal development is sensitive to maternal nutritional, chemical, and environmental stressors. In utero exposures may compromise early developmental processes and predispose the fetus to adverse health risks later in life, based on the Developmental Origins of Health and Disease (DOHaD) hypothesis2. Epidemiologic studies have also demonstrated that prenatal exposure to BPA, phthalates, and polyfluoroalkyl chemicals may be associated with fetal growth restriction or small for gestational age babies27, both of which are risk factors for adult onset disease30. However, the literature is mixed, with some studies demonstrating no effect of EDC exposure on birth outcomes or anthropometric measures32.Pregnancy is a clinically relevant susceptibility period for investigating EDC exposures, due to the growing body of evidence that EDCs modulate physiologic processes in both exposed individuals and their offspring34. EDCs may exert physiologic effects at low doses, and modulation of hormonal pathways occurs in a non-monotonic manner for substances such as BPA, dioxin, lead, and cadmium36. While most EDCs have been studied in isolation, there is increasing interest in the cumulative impact of EDC exposures, as humans are exposed to a multitude of EDCs at varying doses which may have additive, synergistic, or negative biologic effects39.Biologically, EDCs may mimic or interfere with estrogenic, androgenic, glucocorticoid, thyroid, and insulin signaling pathways, leading to a variety of downstream, tissue-specific effects41. Further, there is evidence that EDCs such as BPA and polycyclic aromatic hydrocarbons, and metals such as arsenic, cadmium, lead, and mercury, may modulate the immune system and alter inflammatory cytokine milieu to favor a pro-inflammatory state44. Meanwhile, findings from other pregnancy cohorts have demonstrated that urinary paraben and phenol measures may affect circulating markers of inflammation including IL-6, IL-10, CRP, and TNF-α46. Importantly, aberrant maternal inflammatory pathways have been linked to adverse pregnancy outcomes including pregnancy loss, preterm labor, preeclampsia, and fetal growth restriction50. Epidemiologic and animal studies indicate that such alterations induced in the maternal environment by EDCs may alter the developmental trajectory of the developing fetus via epigenetic modifications51.EDC exposures in pregnancy have been associated with changes in the gestational endocrine milieu, including altered levels of sex steroidsThe purpose of our pilot study is to use the Michigan Mother-Infant Pairs (MMIP) birth cohort to assess the impact of phenols, phthalates, and metals, both individually and in combination, on the maternal and neonatal inflammasome. Using maternal urinary measures of environmental toxicants during the first trimester, a sensitive period in fetal development, we correlated exposures to plasma levels of inflammatory cytokines in both maternal (at two time points) and neonatal umbilical cord blood. Exposures and inflammatory cytokine levels were then correlated with birth outcomes, namely infant birth weight and gestational age at delivery.53.The MMIP project is an ongoing birth cohort study (2010-present) at the University of Michigan (UM). Pregnant women are recruited into MMIP at their first prenatal appointment between 8 and 14 weeks of gestation. Those eligible for MMIP participation are between 18 and 42 years old, have a naturally conceived singleton pregnancy, and intend to deliver at UM. Pregnancy dating is determined by the obstetric provider, either based on last menstrual period or ultrasound. As of October 2018, 289 women in the MMIP cohort have delivered. Other studies involving subsets of MMIP participants have been published previouslyA subset of fifty-six women, recruited between November 2012 and May 2015, were selected for the specific analyses described here. All 56 women met additional inclusion criteria as follows: 1. had available biospecimens from mother and infant, at all relevant time points, and 2. had provided complete demographic and health information at their initial study visit. Study procedures were approved by the UM Institutional Review Board, and all participants provided written informed consent. All research was performed in accordance with relevant guidelines and regulations.Spot urine and venous blood samples were collected from participants at their first prenatal appointment between 8 and 14 weeks of gestation, and upon arrival to the hospital for delivery, prior to IV placement. Urine samples were collected in polypropylene containers before being transferred to glass vials. All samples were stored at −80 degrees Celsius until analysis.In total, 41 exposure measures were quantified per subject. These 41 exposures comprise a wide range of commonly encountered environmental toxicants and were chosen due to their heterogeneity.et al., 201853. Briefly, using maternal urine samples collected between 8 and 14 weeks’ gestation, 12 phthalate metabolites and 12 phenol metabolites were quantified via isotope dilution liquid chromatography and tandem mass spectrometry.Details on phthalate and phenol exposure assessment have been previously published by Montrose Phthalate analytes included metabolites of diethylhexyl phthalates (DEHP): mono (2-ethyl-5-carboxylpentyl) phthalate (MECPP), mono (2-ethyl-5-hydroxyhexyl) phthalate (MEHHP), mono (2-ethylhexyl) phthalate (MEHP), and mono (2-ethyl-5-oxohexyl) phthalate (MEOHP); and dibutyl phthalates (DBP): mono-isobutyl phthalate (MIBP) and mono n-butyl phthalate (MnBP). Concentrations of mono-benzyl phthalate (MBzP), mono-carboxy isononyl phthalate (mCINP), mono (3-carboxypropyl) phthalate (MCPP), mono (6-COOH-2-methylheptyl) phthalate (MCOMHP), monoethyl phthalate (MEP), and mono-isononyl phthalate (mINP) were also assessed.Phenol analytes included parabens , bisphenols , 2,4 and 2,5-dichlorophenol , benzophenone-3 (BP3), triclocarban (TCC), and triclosan (TCS).In first trimester urine samples, 17 heavy metals and metalloids were also quantified via isotope chromatography plasma tandem mass spectrometry. These included arsenic (As), barium (Ba), beryllium (Be), cadmium (Cd), chromium (Cr), copper (Cu), mercury (Hg), manganese (Mn), molybdenum (Mo), nickel (Ni), lead (Pb), selenium (Se), tin (Sn), thallium (Tl), uranium (U), tungsten (W), and zinc (Zn).54.Concentrations below the limit of detection (LOD) were assigned a value of LOD/√2. Specific gravity of maternal urine samples to indicate urine dilution was measured via a digital handheld device . Detection limit and coefficient of variation for the EDC measures has been previously addressedMaternal venous blood samples at two time points , and neonatal umbilical cord blood samples were obtained, divided into aliquots, and stored at −80 degrees Celsius. Cord blood samples were of mixed arterial and venous origin, and obtained and stored within thirty minutes of delivery. Maternal and neonatal plasma samples were then analyzed at the clinical chemistry laboratory at the Michigan Diabetes Research Center  for the following inflammatory cytokines: granulocyte macrophage colony-stimulating factor (GM-CSF), interferon (IFN)-γ, monocyte chemotactic protein (MCP)-1, MCP-3, macrophage inflammatory protein (MIP)-1α and MIP-1β, tumor necrosis-factor (TNF)-α, vascular endothelial growth factor (VEGF), interleukin (IL)-1β, IL-6, IL-8, and IL-17α. These 12 cytokines were chosen due to their various roles in immune function. Cytokines were analyzed in duplicates and the average measurement recorded for data analysis. Cytokine levels below the limit of detection (LOD) were replaced by LOD/√2. LOD for individual cytokines was: GM-CSF 0.43 pg/ml, IFN-γ 0.12 pg/ml, MCP-1 1.10 pg/ml, MCP-3 0.24 pg/ml, MIP-1α 1.14 pg/ml, MIP-1β 0.22 pg/ml, TNFα 0.49 pg/ml, VEGF 0.48 pg/ml, IL-1β 0.06 pg/ml, IL-6 0.4 pg/ml, IL-8 0.17 pg/ml, and IL-17α 0.08 pg/ml. Coefficients of variation for individual cytokines were: GM-CSF 9.1%, IFN-γ 6.0%, MCP-1 5.1%, MCP-3 6.3%, MIP-1α 6.4%, MIP-1β 6.5%, TNFα 7.8%, VEGF 18.9%, IL-1β 7.4%, IL-6 10.4%, IL-8 6.5%, and IL-17α 10.5%.Details on birth outcomes were abstracted from the medical record. Clinical outcomes assessed included estimated gestational age at delivery , mode of delivery, infant sex, and birth weight. Details on possible confounding variables were provided by subjects at their initial study visit and confirmed in the medical record. Possible confounders included pre-pregnancy body mass index (BMI), maternal age, and history of smoking. Because only two women in our cohort developed gestational hypertension, and only one developed gestational diabetes, pregnancy conditions and complications were not included as confounders.To address the issue of measurements falling below the LOD, rank-based methods with ties were applied for correlation and regression analysis that is robust to either fixing truncated measurements at their upper bound or replacing them with LOD/√2.Spearman correlation coefficients were calculated to determine the association between first trimester exposures and inflammatory cytokine markers at two time points (first trimester and delivery) in mothers and at delivery for newborns. EDC measurements were first adjusted for urine specific gravity and then log-transformed prior to correlation analysis. Each of the 12 inflammatory cytokines was tested for correlation with each of the 41 exposures, and FDR Benjamini–Hochberg (BH) correction was employed to account for multiplicity in the hypothesis testing, using FDR adjusted p-value 0.1 as the cutoff of signal calling. To account for confounding variables, including pre-pregnancy BMI, maternal age, history of smoking, mode of delivery , infant sex, and gestational age at delivery, linear regression was used to examine the association between log-transformed exposures and inflammatory markers. Linear regression was also used to test the association between inflammatory cytokine markers and two birth outcomes, infant gestational age and birthweight. For regression analyses, cytokine values were transformed using inverse normal transformation and birth outcomes were transformed using standard normal transformation. Birthweight z-score was calculated from birthweight in grams, and used as the outcome variable.55. This resulted in 11 principal components (PC), each representing one type of linear combination of EDC exposure measures, collectively explaining 80% of the total variation in the data. Linear regression was used to evaluate the association between the PCA-weighted EDC exposures with maternal and fetal inflammatory cytokines, adjusted for confounding variables. PCA was also used to investigate the effect of EDC exposure mixtures on infant birth weight and gestational age at delivery. In the models with PC variables, the same inverse normal transformed cytokines and z-scores of birthweight were used, and p < 0.05 was considered significant. Statistical analysis was completed in R Software, version 3.0.To investigate the effect of exposure mixtures on maternal and fetal inflammatory status, we conducted principal component analysis (PCA) on the 41 natural-log-transformed exposure variables. PCA was used in order to simplify the exposure data into fewer dimensions, and to summarize the data into a smaller number of principal componentsIn addition to the cross-sectional data analysis of inflammatory cytokine measurements at each time point, we conducted a longitudinal analysis using a Linear Mixed Model (LMM) with random intercepts to account for temporal association between repeatedly measured concentrations of maternal cytokine levels in first trimester and at term. The LMM model, with interaction terms between time and PCA-weighted EDC exposures, analyzed both main effects of the PC-weighted EDC exposure and time, and more importantly how the association effect between the exposures and cytokines may have shifted from baseline to term.This study was approved by the University of Michigan IRB Committee.54. In general, women displayed an average of 30 detected EDC exposures during the first trimester, including a mean of 12.8 metals, 10.5 phthalate metabolites, and 7.7 phenols54.Subjects in the current analysis represent 56 mother-infant dyads in the MMIP cohort. Sample characteristics and infant birth outcomes are outlined in Table First trimester measures of several individual EDCs, particularly metals and phthalates, were independently associated with maternal first trimester and term inflammatory markers using Spearman correlation coefficients Figs  and 2. PWhen adjusted for confounders using linear regression, individual EDCs were significantly associated with maternal IL-6 in the first trimester, and maternal MCP-1 and IL-17 at delivery. Table Several positive correlations between inflammatory markers within each time point  were identified  longitudinal analysis, there were statistically significant interactions between time of maternal inflammatory measurement and seven of the PC groupings. These interactions indicate an altered association from baseline to term, which are not captured by the cross-sectional analyses conducted by linear regression. Table Finally, we performed comparative analyses between several subgroups, using mean concentrations (95% CI) of EDC measures and inflammatory markers: maternal obese vs. non-obese, primiparous vs. multiparous, and neonatal sex . However, no significant differences were identified (data not shown).15. Pregnant women in the MMIP cohort demonstrated lower urinary levels of many exposures, such as Hg  and BPA 56. However, some exposure measures appeared to be greater in MMIP subjects, such as TCS (geometric mean 14.5 ug/L [8.28–25.40] vs 50th% percentile 8.2 ug/L in NHANES). These wide ranges of values, both within and between cohorts, may reflect differential exposure due to geography, socioeconomic status, occupation, and other environmental factors.Women in this Michigan-based birth cohort demonstrated detectable exposure to an average of 30 chemicals and metals, out of 41 possible that were measured, in the first trimester. These results are comparable to population studies such as NHANES, which demonstrated a median detectable exposure of pregnant women to 50 chemical analytes, out of 71 possible exposuresFurthermore, in this sample of pregnant women and their full-term newborns, inflammatory marker levels in early gestation and the peripartum period were differentially associated with (1) individual exposures and (2) exposure mixtures. These findings suggest that not only do these environmental exposures alter the maternal-fetal inflammasome throughout pregnancy, but that the combination of exposures is of particular importance. Furthermore, several positive associations between individual inflammatory measures and two birth outcomes, infant birth weight and gestational age, were also evident.58. For the 12 inflammatory markers measured in our study, our subjects (excluding outliers) demonstrated similar cytokine levels in the first trimester and at term  and a similar trend between the phthalate metabolite DEHP with IL-662. In contrast, in our cohort of Caucasian women, neither TCS nor DEHP metabolites were found to be individually associated with IL-6. These differences may relate to differences in the timing of EDC and inflammatory measures, in addition to ethnic differences. Another pregnancy cohort in Boston demonstrated a link between maternal BPA and IL-6 levels at various points in pregnancy, including the first trimester63. This finding was not observed in our MMIP population, possibly due to differences in patient demographics, regional exposures, and/or statistical power.Our findings also suggest that maternal exposure to EDCs may contribute to perturbations in antepartum and peripartum cytokine levels in not only the pregnant mother, but also the newborn. A unique aspect of our study is that maternal inflammatory markers were related to EDCs measured in early pregnancy (8–14 weeks’ gestation), a critical window of fetal and placental development. In relation to individual exposures, increased first trimester exposure to metals and phthalates was found to be positively associated with MCP-3 and IL-8 levels, while increased exposure to metals was negatively associated with MCP-1. However, when adjusted for confounders , first trimester IL-6 was the only cytokine with a significant association to environmental EDCs . One published study involving a Puerto Rican pregnancy cohort64. Its relevance to maternal-fetal outcomes warrants further investigation.At term, just prior to delivery, we found that maternal MCP-3 maintained positive relationships with first trimester measures of metals and phthalates. Interestingly, the correlation between MCP-3 and first trimester metal exposures was also demonstrated in umbilical cord blood, although this correlation disappeared after adjusting for confounding variables. MCP-3 attracts monocytes and macrophages but its specific role in pregnancy has not been defined65. To our knowledge, ours is the first study to directly compare neonatal cord blood inflammation with early trimester maternal EDC measures. Assessing maternal exposure in early pregnancy, and not at the time of delivery, allows investigation of the maternal exposome during important periods in fetal development, which is a unique feature of our study.There is limited data on the relationship between EDC exposure with cord blood markers, despite the fact that cord blood samples are relatively easy to obtain. Measures of cord blood exposures may be an effective proxy for maternal exposures during pregnancy, as chemicals in maternal blood are often also detected in cord blood67. Notably, comparison of individual EDCs to various groupings of EDC mixtures demonstrated different associations with cytokine levels. The principal component (PC) group 11 in our dataset provides an excellent example of this. As evidenced in Tables 68. In addition, the observed longitudinal changes of inflammatory biomarkers from first trimester to term, in relation to EDC mixtures, suggest that EDC exposures may alter the physiologic cytokine shifts which occur throughout pregnancy. However, the exact contribution of these combined exposures to longitudinal cytokine levels, and the clinical implications of these findings, remain unknown at this time.Use of principal component analysis (PCA) identified several notable associations between EDC mixtures and maternal/neonatal cytokine levels. These findings contribute to a growing body of literature on the impact of chemical mixtures on human health67. Some chemical combinations are more prominent than others depending upon socioeconomic, occupational, dietary, behavioral, and geographic factors of the study population70. More research is needed to identify the most prevalent exposure mixtures among specific populations, and their cumulative impact on physiologic processes66. Although the EDC exposure assessment undertaken in this study was comprehensive, with urinary measures of 41 chemicals, in reality humans are exposed to innumerable chemicals via various routes of administration including oral, inhaled, dermal, and ingested. The complex nature of these “real life” exposures, and the challenges of quantifying them, is demonstrated in assessing environmental exposure risks in the agriculture, oil and gas industries72. In addition, EDCs may modulate downstream molecular and cellular pathways in a dose-independent, non-monotonic, sex-specific manner, and may be influenced by other factors such as genetics, nutrition and stress2, which may be difficult to account for.In practice, there are methodologic challenges to studying mixture interactions and their physiologic effects76, some studies demonstrate no association78. In our 56 subjects, we only observed associations between birth weight and two exposures, MCPP (positive) and BPS (negative)54. However, associations between inflammatory cytokine levels and birth outcomes were evident. Our findings revealed higher levels of first trimester MIP-1α, IL-8, and TNF-α to be associated with an increase in infant birth weight, while first trimester IL-6 was found to be associated with an increase in gestational age. In contrast, no significant associations were found between the inflammatory milieu at the time of delivery with birth outcomes. These observations have important translational value. Aberrant inflammation in pregnancy has been associated with a spectrum of adverse outcomes related to implantation, placentation, and fetal growth, including implantation failure, recurrent pregnancy loss, preeclampsia, intrauterine growth restriction, and preterm birth83. A link with inflammation is particularly well-established in the preterm population86. In our cohort, all newborns were delivered at term, which may explain why an association between gestational age and inflammation at delivery was not observed. Further, all newborns in our cohort were of normal birth weight. Although in the original DOHaD hypothesis87, low birth weight was the primary risk factor for adult-onset disease, there is increasing evidence that developmental programming may occur in a variety of pathways, including inflammatory, endocrine, epigenetic, nutritional, oxidative stress, and/or placental91. Our findings of an altered inflammatory milieu in utero may still confer long-term health programming effects on offspring, even in the setting of normal birth weight.While several epidemiologic studies demonstrate an association between EDC burden and low birth weight in humans92, we demonstrate that reproductive age women are widely exposed to multiple classes of environmental chemicals. Because gestational EDC exposures may have long-term health implications for adult offspring through developmental programming, potentially due to inflammatory mechanisms, this particular area of research has significant clinical and public health implications for women of childbearing age93.To our knowledge, this is the first study to support an association between EDC mixtures and alterations in the maternal and neonatal inflammatory environment. Similar to other studies94, but due to small sample size, male and female newborns were not analyzed separately. Limitations of our study design include the fact that we did not account for hemodilution or hemoconcentration of plasma samples. Furthermore, EDC exposures were only measured at a single time point in pregnancy. Exposure measures may vary throughout pregnancy, and repeated measures of EDC levels over time may better represent average prenatal exposure96. By only measuring EDCs in the first trimester, we were unable to assess the impact of second or third trimester exposures on inflammatory processes. Other possible confounders, which were not addressed by our statistical design, include pre-pregnancy maternal comorbidities, maternal weight gain in pregnancy, parity, medication use, vitamin or supplement intake during pregnancy, and whether labor was initiated spontaneously or via induction. Finally, although our pilot study is unable to assess mechanisms or establish causality, our findings highlight the potential association of circulating inflammatory markers in pregnancy with EDC mixtures, which warrants additional research.Strengths of our study include our comprehensive assessment of the association between 41 first trimester exposures, with 12 inflammatory cytokines in early pregnancy and the peripartum period. Recruitment of pregnant women at their first prenatal appointment allowed assessment of the maternal inflammasome in the first trimester of pregnancy, when fetal development is particularly sensitive to insults. Our study is limited by a small sample size and a non-diverse patient population. However, our subject homogeneity does limit the potential for additional confounding variables including race, education, and socioeconomic status. As previously mentioned, EDCs may exert their effects in a sex-specific mannerIn conclusion, inflammatory cytokine levels throughout pregnancy and at delivery may be explained by unique mixtures of first trimester environmental toxicant exposures. More research is needed to validate these findings on a larger scale and to assess their impact on a wider range of pregnancy and birth outcomes.Supplemental Figures/Tables


Evidence suggests the possible impact of ambient high temperature on fetal growth and birth outcomes. However, little is known about the relative impact of exposure to heat and cold and the possible vulnerable window during pregnancy.10, SO2, NO2, and O3, were obtained from relevant government agencies. This study was to examine the associations of maternal exposure to ambient temperature  with the duration of gestation and birth weight.Data on a total of 237,585 pregnant women from January 1st, 2001 to December 31st, 2010 were acquired from the Queensland Health, Australia. Daily data on meteorological factors, including ambient temperature, relative humidity, barometric pressure, and air pollutants, such as PMA J-shaped association between minimum temperature at conception and duration of gestation was observed after adjusting for seasonality and other confounders. Compared to women who were exposed to the minimum temperature of 15–20 °C in the first gestational week, exposure to the minimum temperature of > 20 °C significantly increased the duration of gestation by 0.029 weeks . A cumulative effect was found when exposure across the first four weeks was examined. There was an inverted U-shaped relationship between minimum temperature at delivery and the duration of gestation. Compared to women exposed to 15–20 °C, exposure to minimum temperature of > 20 °C and ≤ 10 °C was associated with a shortened gestation by 0.030 weeks  and 0.018 weeks , respectively. By contrast, an inverse relationship between maximum temperature and birth weight was observed. Compared to exposure to the maximum temperature of > 30 °C in the last week of pregnancy, maternal exposure to 20–25 °C and < 20 °C significantly increased birth weight by 0.011 kg  and 0.018 kg , respectively. Similarly, a mild cumulative effect was observed when maximum temperature exposure across the four weeks before delivery was evaluated.The finding emphasized the importance of keeping an optimal temperature range during pregnancy for reducing the risk of preterm birth and low birthweight.The online version of this article (10.1186/s12884-018-2100-y) contains supplementary material, which is available to authorized users. Climate has frequently been changing throughout the Earth’s history. However, the recent pace of warming far exceeds that of any previous warming episode in the past 10,000 years , 3. GlobPrevious studies have explored the short-term and long-term effects of maternal exposure to heat stress on preterm birth, stillbirth, and birth weight –17. In gThe effect of ambient temperature on birth outcomes may depend on the stage of gestation, with a possible susceptible period during pregnancy , 21. ForBased on above, we speculated the possibility that high temperature and cold condition could have varied health effect, and, meanwhile, the effect may depended on different exposure period of gestation. This study focused on two important birth outcomes, duration of gestation and birth weight, aimed to address two issues: 1) what is the overall impact of both hot and cold temperatures on duration of gestation and birth weight? 2) is there any possible susceptible period to temperature exposure for different pregnant outcomes?Participants in this study were all the singleton births born in Brisbane, Australia, between January 1st, 2001 and December 31st, 2010. Data were collected from the Perinatal Data Collection Unit (PDCU) of the Queensland Health Statistics Centre. All births were recorded according to the following criteria: 1) at least 20 weeks of gestation; or 2) at least 400 g in birth weight; 3) born in Queensland including all public hospitals, private hospitals, and private midwifery and medical practices.In this study only live birth was eligible. Among all live births, those whose last menstrual date was earlier than January 1st 2001 were excluded to match our environmental exposure records that covered from January 1st 2001 to December 31st 2010. The final sample consisted of 237,585 singleton live births. The study profile and participants selection process are shown in Fig. The following variables were collected based on standardized maternal medical records: duration of gestation, date of last menstrual period, date of birth, gender of baby, birth weight, mode of labor onset , maternal age group, marital status , Indigenous status, and parity.Meteorological factors, including daily maximum and minimum temperatures, relative humidity at 9:00 am and 15:00 pm, and ambient barometric pressure at 9:00 am and 15:00 pm, were recorded in nine monitoring stations in Brisbane during the study period.3), nitrogen dioxide (NO2), and sulfur dioxide (SO2), were recorded in five monitoring stations in Brisbane.Air pollutants, including particulate matter with a diameter < 10 μm (PM10), ozone  based on the original daily data. Since the date of conception and birth varied among different pregnant women, we, then, fitted weekly data of these environment indicators to each pregnant woman, using their mean levels in the first/last four weeks of the corresponding gestation. To compare the possible acute and cumulative effects of temperature exposure, the exposures in the first one gestational week vs. in the first four gestational weeks and in the last gestational week vs. in the last four gestational weeks were examined, respectively, in the subsequent analyses. Descriptive statistics for meteorological and pollutant indicators at conception/delivery are shown in . Generalized cross validation was used to automatically select the degree of smoothing for spines. Based on the distribution of temperature and overall effect of temperature on duration of gestation/birth weight in GAM, maximum temperature was recorded into following four groups as > 30 °C, 25–30 °C, 20–25 °C, and ≤ 20 °C, while minimum temperature was grouped as > 20 °C, 15–20 °C, 10–15 °C, and ≤ 10 °C.10, O3, NO2, and SO2) around conception and delivery, were further controlled. Accumulating studies documented that other seasonal factors , independent of ambient temperature, may affect fetal development and birth outcomes [a), and temperature around conception were simultaneously taken into account when examining the health effect of temperature exposure at delivery (Model 3b).Generalized linear regression models were further applied to estimate the crude and adjusted associations between maximum/minimum temperature groups  and duration of gestation/birth weight . Adjustments were made following a three-step procedure. Model 1 was adjusted for all maternal and perinatal variables (as listed in Table outcomes –24. Therwww.r-project.org). In the presentation of the results, the statistical significance was set at p value < 0.05 (two tailed).All analyses were performed using the Statistical Analysis System (SAS) for Windows, version 9.2  and R version 2.15.1 , and the mean birth weight was 3.42 kg . Table p < 0.001). The smoothed plots of nonlinear associations of duration of gestation with minimum temperature at the first/last gestational week are shown in Fig. , both p < 0.001). By contrast, approximate linear trend or wave curves, without clear correlates, were observed between maximum temperature, either at concept or at delivery, with duration of gestation .The association between ambient temperature and duration of gestation/birth weight was explored by penalized splines in generalized additive models. We observed a J-shaped association of duration of gestation with both minimum temperatures at conception and at delivery (both p < 0.05). Associations of birth weight with maximum temperature at concept or minimum temperature at concept/delivery were also explored. Generally, there were wavy curves without clear correlates (not shown).The smoothed plots of association between birth weight and maximum temperature at the last gestational week is shown in Fig. Table p = 0.006) and 0.049 weeks (p < 0.001).Table p < 0.001), 0.030 weeks (p = 0.007) and 0.018 weeks (p = 0.041) of the shortened duration of gestation, respectively.Minimum temperature at delivery was also related to the duration of gestation. Compared to minimum temperature exposure of 15–20 °C in the last week, > 20 °C and 10–15 °C could significantly shorten the duration of gestation, respectively. Similarly, a mild cumulative effect was found. During the process of three-step adjustment, the associations generally maintained. In the final full model (Model 3), exposure to ambient temperature > 20 °C in the last gestational week, as well as > 20 °C and ≤ 10 °C in the last four gestational weeks, were associated with a decreased gestation age, when compared to 15–20 °C, and there were 0.042 weeks  and ≥ 37 weeks (full-term birth) to examine the relationship between exposure to ambient temperature and risks of preterm birth  and 0.018 kg (p = 0.024), respectively. A slightly cumulative effect was identified.Table Results of this study indicate a complex spectrum about the association between ambient temperature exposure and birth outcomes. For the first time, we reported that the duration of gestation seemed to be sensitive to maternal exposure to minimum temperature, while birth weight was susceptible to maximum temperature exposure.Most previous studies have focused on the relationship between high temperature and birth outcomes –17. HoweTo our knowledge, this is the first study that has investigated the relation between ambient low temperature and duration of gestation. However, two recent studies examined the impact of ambient low temperature on preterm birth. The one in Uppsala, Sweden set up a retrospective birth cohort among almost 14,000 deliveries from 1915 to 1929 and, in which, it was found that extreme cold exposure adversely affected preterm birth . HoweverAn almost linear relationship between maximum temperature and birth weight was observed. Compared to exposure to higher than 30 °C at the last week of pregnancy, birth weight significantly increased by 11 g and 18 g for those who exposed to 20–25 °C, and < 20 °C, respectively. Similarly, a mild cumulative effect was observed when maximum temperature exposure across the four weeks before delivery was evaluated. By contrast, minimum temperature was not found to be related to birth weight. The findings indicated that the period before delivery could be a susceptible window for fetal weight growth when high ambient temperature exposure exists.r = − 0.22, P < 0.01) [r = − 0.59, P < 0.01) between heat index and birth weight, where it was found that a one unit increase in heat index was associated with 2.7% decrease in birth weight [Among the previous studies on the topic of maternal temperature exposure and birth weight, there was a decrease in birth weight associated with increasing temperature at delivery , 28, 32. < 0.01) . An ecolh weight . The resh weight . Our resh weight , this fih weight . In addih weight . Pregnanh weight –8. The iA few studies have explored the possible impact of cold ambient temperature on birth weight , 25, 36,It is crucial to identify the possible vulnerable exposure window for a specific birth outcome. However, a methodological challenge arises when dealing with preterm birth, when pregnancy doesn’t consist of three full trimesters. The common compromised approach was to focus on exposure during the period before delivery , 26–30 oSeveral limitations should be acknowledged in interpreting the results. The principal limitation lies in ecological assessment of exposure, both for meteorological indicators and air pollutants. Personal exposure may be modified or attenuated by the duration spent indoors. In addition, the present study relied on the mothers’ report of the date of the last menstrual period to determine gestational age at birth, gestational weeks were used as an analytic scale in this study. A 24 h time frame may be not sufficiently accurate to quantify the difference of several hours, but this bias is likely to be non-differential and to lead the estimates towards null hypothesis. Moreover, as a population-based study, even a small reduction in mean gestational age can lead to a considerable increase in the occurrence of preterm birth. Finally, although we controlled several possible confounding factors, other factors, such as health condition and lifestyle behaviors, may be significantly associated with birth outcomes.This study provided new insights and enriched the understanding of the relationship between ambient temperature and birth outcomes through the following findings: (I) There were different pregnancy periods vulnerable to ambient temperature exposure for different birth outcomes. For the duration of gestation, both the early and late pregnant periods were important; while for birth weight, the late pregnant period was more important than the early period. (II) Duration of gestation seemed to be sensitive to minimum temperature, while birth weight was susceptible to maximum temperature. (III) A J-shaped and an inverted U-shaped associations were observed between the duration of gestation and maternal minimum temperature exposure at conception and delivery, respectively. Meanwhile, there appeared to be an almost linear relationship between maternal exposure to maximum temperature and birth weight.Additional file 1:Table S1. Summary distribution of meteorological and pollutant exposure of 237,585 births in Brisbane, 2000–2010. Summary distribution of meteorological and pollutant exposure. (DOCX 27 kb)Additional file 2:Figure S1. The associations between ambient minimum temperature and gestational age after adjustment for maternal and perinatal factors, air pollutants, and meteorological exposure. The associations between ambient minimum temperature and gestational age. (PDF 115 kb)Additional file 3:Figure S2. The associations between ambient maximum temperature and gestational age after adjustment for maternal and perinatal factors, air pollutants, and meteorological exposure. The associations between ambient maximum temperature and gestational age. (PDF 106 kb)Additional file 4:Figure S3. The associations between ambient maximum temperature and gestational age after adjustment for maternal and perinatal factors, air pollutants, and meteorological exposure. The associations between ambient maximum temperature and gestational age. (PDF 109 kb)Additional file 5:Figure S4. The associations between ambient maximum temperature and birth weight after adjustment for maternal and perinatal factors, air pollutants, and meteorological exposure. The associations between ambient maximum temperature and birth weight. (PDF 82 kb)Additional file 6:Table S2. The description of full-term birth and preterm birth by different ambient temperature category in Brisbane, 2000–2010 . The description of full-term birth and preterm birth. (DOCX 23 kb)Additional file 7:Table S3. The associations of ambient temperature at conception/delivery with preterm birth in Brisbane, 2000–2010 . Association of ambient temperature with preterm birth. (DOCX 31 kb)