Heart failure (HF) is a complex and multifactorial syndrome caused by impaired heart function. The high morbidity and mortality of HF cause a heavy burden of illness worldwide. Non-thyroidal illness syndrome (NTIS) refers to aberrant serum thyroid parameters in patients without past thyroid disease. Observational studies have indicated that NTIS is associated with a higher risk of all-cause mortality in HF. This meta-analysis aimed to investigate the association between NTIS and HF prognosis.I2 and Cochran's Q statistics. Sensitivity analysis and publication bias analysis were also conducted.Medline, Embase, Web of Science, and the Cochrane database were searched for any studies reporting an association between NTIS and HF prognosis from inception to 1 July 2022. A meta-analysis was then performed. The quality of studies was assessed using the Newcastle–Ottawa Scale. The heterogeneity of the results was assessed with A total of 626 studies were retrieved, and 18 studies were finally included in the meta-analysis. The results showed that NTIS in HF patients was significantly associated with an increased risk of all-cause mortality and major cardiovascular events (MACE), but not with in-hospital mortality. The stability of the data was validated by the sensitivity analysis. There was no indication of a publication bias in the pooled results for all-cause mortality and MACE.This meta-analysis showed that NTIS was associated with a worse outcome in HF patients. However, the association between NTIS and in-hospital mortality of HF patients requires further investigation. Heart failure (HF) is a complex, multifactorial clinical syndrome characterized by dyspnea and fluid retention caused by impaired systolic or diastolic cardiac function , 2. In 2Non-thyroidal illness syndrome (NTIS) was first discovered in the 1970s and described as abnormal serum thyroid parameters in patients with acute illness but without prior thyroid disease. The most common abnormality is low triiodothyronine T3) levels with normal thyroid-stimulating hormone levels  levels w. FurtherTherefore, we performed this meta-analysis to explore the association between NTIS and the HF prognosis for various endpoints and subgroups. To our knowledge, this is the first meta-analysis to systematically review these associations.https://www.crd.york.ac.uk/PROSPERO).This meta-analysis was performed according to the Preferred Reporting Items for Systematic Review and Meta-Analyses (PRISMA) guideline 2020 . We regi3 syndrome OR low triiodothyronine syndrome OR euthyroid sick syndrome OR ESS). Two reviewers (XYQ and YXZ) screened the titles and abstracts of the retrieved studies. The full texts of relevant studies were further evaluated, and reviews, case reports, animal experiments, and studies without prognosis information were excluded. A third reviewer (YJC) was consulted when there was a difference of opinion between the first two reviewers.We conducted a comprehensive literature search of the Medline, Embase, Web of Science, and Cochrane databases to obtain studies focused on the association between NTIS and the HF prognosis, published from inception to 1 July 2022. The inclusion criteria were as follows: (i) HF patients were included in the study populations; (ii) cohort studies; (iii) studies investigating the association between NTIS and HF prognosis; (iv) study endpoints included all-cause mortality or major cardiovascular events (MACE), or in-hospital mortality; and (v) studies published in English. Our search strategy was therefore: (heart failure OR heart decompensation OR congestive heart failure OR HF) AND , relative risk (RR), HR), with 95% confidence intervals (95% CIs) and adjusted covariates for multivariable analysis.We used the Newcastle–Ottawa Scale (NOS) to evaluate the quality of the eligible cohort studies . The NOSI² and Cochran’s Q statistics were used to assess heterogeneity. A P-value < 0.05 for Q statistic and I² > 50% was considered substantial heterogeneity. The sensitivity analysis was performed by omitting each study to evaluate the consistency of the pooled results. Publication bias was assessed using the funnel plot and Egger’s test. Obvious asymmetry of the funnel plot and a P-value < 0.05 for Egger’s test indicated significant publication bias. The trim and fill method would be used if publication bias existed. Subgroup analyses were conducted on disease onset of HF, NTIS definition, continent of origin of the study population, and source of HRs among different endpoints, that is, whether the HRs were reported or extracted from survival curves. A subgroup analysis according to MACE subtypes was also performed. All statistical analysis was performed using Stata software (Version 14.0). A two-sided P-value<0.05 was considered statistically significant.The RR, OR, or HR were extracted from the original studies. In studies that only gave survival curves, we calculated the crude HR using Engauge Digitizer 4.1 software , 11. Poon = 263), animal studies (n = 10), case reports (n = 21), and reviews (n = 98), were excluded. Forty-two reports were retained for further examination. We excluded four further studies that were conference abstracts or for which we had no access to the full text, five studies that were not focused on HF patients, ten that were cross-sectional studies, and five that did not have data that would enable the determination of effect size. Finally, our meta-analysis included 18 cohort studies that examined the relationship between NTIS and HF prognosis. The workflow is shown in The initial database search identified 656 articles. After eliminating 222 duplicate studies, we screened the titles and abstracts of the remaining records. Studies not relevant to our topic . The pooled ORs also demonstrated a consistent result with high heterogeneity .Of the 12 studies that explored the association between NTIS and all-cause mortality in HF patients , HRs werP < 0.001, I2 = 0; AHF: HR: 2.12, 95% CI: 1.53–2.94, P < 0.001, I2 = 0). For subgroup analysis based on the definition of NTIS, the synthesis of HRs showed that the association between NTIS and all-cause mortality of patients with HF was significant in both groups . Subgroup analysis based on the different continents of origin showed that NTIS was significantly associated with an elevated risk of all-cause mortality in studies from Europe, America, and Asia . Finally, subgroup analysis of the source of the HRs indicated that the association between NTIS and all-cause mortality in HF patients was significant regardless of the HRs origins, that is, reported or extracted from survival curves . However, the heterogeneity of pooled HRs from the survival curve (SC) group increased (I2 = 53.6%).The results of the subgroup analyses of pooled HRs are shown in P < 0.05)  .A sensitivity analysis was performed to assess the stability of the pooled HRs/ORs by omitting each study. The association between NTIS and all-cause mortality of HF patients remained significant after the removal of each study . Further, pooled ORs also confirmed a significant association between NTIS and MACE in HF patients . The heterogeneity of the pooled HRs and ORs was significant (I2 > 50%), so we conducted the following subgroup analyses  but not in AHF patients . For subgroup analysis based on the definition of NTIS, the association between NTIS and MACE in HF patients was significant under both strict and no-strict definition . Subgroup analysis based on the continents of origin demonstrated that NTIS was significantly associated with an elevated risk of MACE in HF patients in studies from Europe and America , but not significant in studies from Asia . Further subgroup analysis on the source of HR indicated the significant association between NTIS and MACE in HF patients with both reported and extracted HRs . For the subgroup analysis of the type of MACE, the result showed that NTIS was significantly associated with an increased risk of MACE for major adverse events (MAE), readmission for HF, and cardiac death . However, the association was not significant in the subgroup of cardiac death and readmission for HF . Additionally, the heterogeneity of subgroups on disease onset, definition of NTIS, continents, and source of HR decreased compared to the pooled result, which suggests the potential source of heterogeneity.The subgroup analysis of disease onset showed that NTIS was significantly associated with a higher risk of MACE in CHF patients  .Sensitivity analysis confirmed the stability of pooled HRs and ORs . The funnel plot and Egger’s test did not reveal a tendency for publication bias (P > 0.05). We did not perform subgroup and sensitivity analysis due to the limited number of studies. The Egger’s test revealed a potential publication bias (P < 0.05), but the use of the trim and fill method did not produce a reversion of the pooled result .Two studies that reported endpoints of in-hospital mortality in HF patients were included in the meta-analysis . The pooIn this meta-analysis, we explored the association between NTIS and both short- and long-term outcomes of HF patients. The results indicate that NTIS is significantly associated with an increased risk of all-cause mortality, and MACE in HF patients, whereas the association with in-hospital mortality was not significant. In addition, the association between NTIS and all-cause mortality remained significant in several subgroups. However, the association between NTIS and MACE was not significant in the subgroup of AHF patients, studies from Asia and MACE defined as cardiac death and readmission to hospital.In contrast to the earlier meta-analysis focusing on the association between NTIS and cardiovascular disease, we discussed NTIS and HF prognosis. Additionally, we analyzed the association between NTIS and multiple outcomes of HF with more studies included. In the original studies, the effect size reporting the association between NTIS and HF prognosis was not uniform. To avoid potential bias, we performed separate meta-analyses for HRs and ORs. The pooled HRs and ORs confirmed that the association between NTIS and all-cause mortality, and MACE in HF patients was significant. There was no heterogeneity in the pooled HRs for all-cause mortality. In the subgroup analysis based on the source of HRs, the heterogeneity increased in the pooled HRs extracted from survival curves, which might be due to the differences in the ethnic and exclusion criteria of the two studies. However, after the exclusion of HRs extracted from survival curves, the result still supports the association of NTIS with HF prognosis. On the other hand, the meta-analysis of MACE revealed significant heterogeneity. However, the heterogeneity among subgroups including disease onset, definition of NTIS, source of HR, and continents of origin, was not significant, which indicated the heterogeneity source. Additionally, the association between NTIS and MACE in HF patients was not significant in several subgroups. The association was not significant in AHF patients, which may be explained by the limited number of studies, or a difference in the disease prognosis profile. A survey by the European Society of Cardiology (ESC), the ESC-HF pilot study survey, revealed that, compared with AHF, CHF had a better medium-term prognosis due to more mature therapies . Further3), by the deiodinase enzyme  α, and calcium ATPase protein of the sarcoplasmic reticulum (SERCA2α), which enhance systolic and diastolic cardiac function. Indirect effects of T3 on the cardiovascular system include the activation of sodium, potassium, and calcium membrane channels, which induces the endothelial nitric oxide production, and the activation of PI3K/AKT pathways to reduce vascular resistance  plays a central role in regulating the cardiovascular system. Thyroxine is converted to the biologically active form of TH, triiodothyronine  and Guangdong-Shenzhen Joint Fund Youth Project (No. 2021A1515111110).


HIF1A rs11549465 and rs11549467, VEGFA rs2010963, and rs833061. The findings suggest that the HIF1A rs11549465T allele is an independent protective factor against necrotizing enterocolitis (NEC), but may increase the risk of diffuse white matter injury (DWMI) in newborns exposed to hypoxia at birth and long-term oxygen supplementation. In addition, the rs11549467A allele was found to be an independent protective factor against respiratory distress syndrome (RDS). No significant associations with VEGFA SNPs were observed. These findings indicate the potential involvement of the hypoxia-inducible pathway in the pathogenesis of complications of prematurity. Studies with larger sample sizes are needed to confirm these results and explore their clinical implications.Excessive oxidative stress resulting from hyperoxia or hypoxia is a recognized risk factor for diseases of prematurity. However, the role of the hypoxia-related pathway in the development of these diseases has not been well studied. Therefore, this study aimed to investigate the association between four functional single nucleotide polymorphisms (SNPs) in the hypoxia-related pathway, and the development of complications of prematurity in relation to perinatal hypoxia. A total of 334 newborns born before or on the 32nd week of gestation were included in the study. The SNPs studied were  Preterm birth is a significant cause of perinatal mortality and morbidity, with an estimated 15 million premature births occurring annually worldwide, accounting for 11% of all births . In PolaThe postnatal transition from a hypoxic intrauterine environment with 20–25 mmHg oxygen tension to an extrauterine normoxic environment of approximately 100 mmHg oxygen tension is an oxidative challenge for newborns ,12. PremVEGFA) is a key activator of hypoxia-induced angiogenesis.Hypoxia-inducible factor-1 (HIF-1) is a crucial regulator of oxygen homeostasis, enabling adaptive responses to hypoxic environments . HIF-1 fHIF1A gene encodes HIF-1α and is located at chromosome 14q21-24. It consists of 15 exons and has over 450 missense variants listed in the Ensembl database. These variants are associated with various predictions of protein function, and two of them, rs11549465  and rs11549467 , located in exon 12, have been extensively studied [HIF1A under either normoxic or hypoxic conditions [HIF1A variants and regulatory elements [VEGFA gene, located at chromosome 6p21.3 and consisting of eight exons exhibiting alternative splicing [VEGFA expression. For example, the rs2010963 (−634G>C) variant is associated with decreased basal promoter activity and reduced protein production, while the rs833061 (−1498C>T) variant has the opposite effect [The  studied ,20. Bothnditions . Anotherelements . The VEGe effect ,26.HIF1A are associated with over 40 different phenotypes and diseases, such as cancer, cardiovascular diseases, and metabolic disorders [HIF1A in the development of prematurity-related diseases has yet to be investigated. On the other hand, previous studies have suggested that VEGFA variants may be associated with female reproductive diseases [VEGFA and the consequences of prematurity, although these have not been consistently replicated [HIF1A SNPs (rs11549465 and rs11549467) and VEGFA SNPs (rs2010963 and rs833061) with the occurrence of NEC, BPD, IVH, RDS, DWMI, or ROP, taking into account the presence of hypoxia at birth. Our research hypothesis is that the genetically determined activity of HIF-1 and VEGF may modulate the risk of developing prematurity-related diseases in infants who have been exposed to perinatal hypoxia.Recent research has identified that genetic variations in isorders ,27. Howediseases ,29,30, aplicated . Thus, tThis study included 334 Caucasian preterm infants who were born between 22 and less than 32 weeks of gestation  criteria) [We analyzed several clinical factors, including gestational age (weeks), birth weight (grams), sex, time of preterm rupture of the fetal bladder (days), mode of delivery , Apgar scores at 1 and 5 min, parameters related to oxygen therapy , and total oxygen supplementation period), as well as blood transfusions. We also collected data on the presence of several diseases of prematurity, including RDS, IVH, BPD, NEC, ROP, DWMI, sepsis, and jaundice of prematurity. We diagnosed RDS, NEC, IVH, BPD, and NEC using criteria that have been described previously , and we 2) ≥0.4 needed to achieve oxygen saturation ≥86% at birth. We chose these criteria, which are less severe than those used to define neonates with birth asphyxia, according to Buonocore et al. [To determine the presence of hypoxia at birth, we assessed whether the newborn fulfilled at least two of the following criteria: acidosis , Apgar score ≤6 at 5 min, and a fraction of inspired oxygen  or from circulating blood lymphocytes using the QIAamp DNA Kit , following the manufacturer’s instructions. Polymorphisms were determined using pre-designed TaqMan SNP genotyping assays  and the ABI 7900HT Fast Real-Time PCR System .χ2 test available at URL http://ihg.gsf.de/cgi-bin/hw/hwa1.pl (accessed on 12 December 2022). Univariate analyses were used to compare the demographic parameters of the study groups. Quantitative parameters were compared using the t-test or Mann–Whitney U test, and qualitative parameters using the χ2 test or Fisher’s test. The Shapiro–Wilk test was used to test the normality of the distribution. Odds ratios (ORs) and 95% confidence intervals (95%CI) were calculated for quantitative parameters. Multivariate analysis was performed using logistic regression analysis. The Bonferroni correction was used for multiple comparisons (padjusted). Statistical analyses were carried out using Statistica v10.0 software and GraphPad Prism v6 software. Quanto software was used for power analysis. Differences were considered significant when p < 0.05.Genotype frequencies were assessed for Hardy–Weinberg equilibrium using the 2 ≥ 0.4), treatment with surfactant, resuscitation, mechanical ventilation, and prolonged oxygen supply. Hypoxia at birth was also identified as a risk factor for several complications, including NEC, BPD, IVH, RDS, ROP, and DWMI, as well as death .HIF1A and VEGFA variants studied, only the HIF1A polymorphisms were found to be associated with prematurity comorbidities. Protective effects against NEC and RDS were observed for the rs11549465T and rs11549467A alleles, respectively. The rs11549465T allele was found also to be a risk factor for IVH and proliferative ROP as well as DWMI, but only in infants exposed to hypoxia at birth .The results of the univariate analyses are shown in HIF1A genotype, hypoxia at birth, and the development of DWMI indicated a gene–environment interaction. Coexistence of carrying the rs11549465T allele and hypoxia resulted in a 5.2-fold increase in the risk of DWMI , compared to the risk of this disease in infants with the rs11549465CC genotype who were not exposed to hypoxia at birth. The observed OR value indicates an amplification of the combined effect over the additive effect of individual factors.The relationship between the HIF1A rs11549465T allele, hypoxia at birth, and prolonged oxygen supplementation increased the risk of DWMI by 7.4 times , while the rs11549467A allele was a protective factor for RDS . In the analysis that included only significant risk factors, the risk of this disease related to coexistence of genotype and hypoxia was increased by 4.5 times . No independent association was found between the HIF1A genotype and the development of IVH and ROP upon multivariate analysis. Low gestational age and birth weight were important risk factors for this condition.The multivariate statistical analysis considered the HIF1A and VEGFA gene variants, which act as a proxy for exposure to increased hypoxia-inducible pathway activity and angiogenesis stimulation, and the occurrence of comorbidities  in newborns born on or before the 32nd week of gestation. Our findings suggest that the HIF1A rs11549465T allele is an independent protective factor against the occurrence of NEC, but it may also increase the risk of DWMI in newborns exposed to hypoxia at birth and long-term oxygen supplementation. Additionally, the rs11549467A allele was an independent protective factor against RDS. We confirmed the role of recognized risk factors such as gestational age, birth weight, and prolonged oxygen supply in the development of prematurity-related diseases. However, we found no associations between the VEGFA gene variants and these diseases.Hypoxia is a trigger of oxidative stress and a significant factor contributing to the pathogenesis of prematurity-related diseases. However, it remains uncertain whether this relationship is causative. In our study, we examined the associations between The interrelationship between gestational age, birth weight, and neonatal morbidity is well documented. The incidence of severe respiratory problems and other morbidities decreases with increasing gestational age and birth weight. The prevalence of prematurity-related diseases reported in the literature varies between European countries and the USA. For example, the incidence of NEC ranges from 2–13%, ROP from 23–52%, BPD from 10–89%, IVH from 31–36%, RDS from 7–50%, and DWMI from 8–20% [The pathogenesis of all these diseases is linked to organ immaturity, insufficient immune and antioxidant systems, environmental factors such as infections and changes in oxygen concentrations, and genetic risk factors ,35,36,37HIF1A genotype in determining susceptibility to birth hypoxia-related DWMI occurrence.According to the literature, the strongest effect of hypoxia at birth can be expected in the case of DWMI, since the central nervous system (CNS) is highly sensitive to oxygen deficiency. During gestation, the development of the CNS is stimulated by exposure to physiological hypoxia, under which HIF-1α is stabilized and plays a crucial role in regulating neural development . PrematuHIF1A alleles and the occurrence of NEC and RDS appears to involve a complex pathomechanism that encompasses various factors associated with fetal development and the activation of the inflammatory cascade by HIF-1α. Animal models of NEC suggest that altered vascular development and tone, which are components of intestinal immaturity, coupled with early exposure to cold stress and a non-milk diet, contribute to the disease [HIF1A SNPs may affect fetal intestinal development and modulate the in utero environment by activating proinflammatory cytokines. Previous studies have linked higher concentrations of HIF-1α in amniotic fluid with a shorter amniocentesis-to-delivery interval and increased levels of IL1α, IL6, and TNFα, indicating the possibility of HIF-1α contributing to the inflammatory cascade in complicated pregnancies [The relationship between  disease . HIF1A Sgnancies . Howevergnancies .The opposite role of high HIF-1 activity in various diseases and conditions is a common observation. Genetically determined high activity of HIF-1, which has proangiogenic effects, benefits adaptation to hypoxia in high altitudes, ischemia related to cardiovascular disease, premature coronary artery disease, and pregnancy complicated by preeclampsia ,45,46,47VEGFA SNPs and the outcomes of interest. However, recent studies suggest that SNPs in this gene other than the ones examined in this analysis may have a greater impact on gene expression and protein levels [On the other hand, high HIF-1 activity may be unfavorable in pathological angiogenesis, accompanying the development of solid tumors and inflammatory diseases, which may also apply to ROP. Experimental studies have demonstrated that pharmacological HIF-1 inhibition can prevent retinal neovascularization, leading to improved visual function in murine oxygen-induced ROP . Anti-VEn levels . TherefoHIF1A SNPs and RDS and NEC. Additionally, the results obtained met the criteria of the Bonferroni correction for multiple comparisons. The other limitation is the lack of data to assess the impact of the SNPs studied on the outcome of ROP treatment and the advancement of IVH. The latter may have contributed to the overrepresentation of IVH in the study population (57.5%) as well as the underestimation of the effect of the HIF1A genotype on IVH. As shown in our previous study, in a smaller group of cases in which any stage of IVH (grade I–IV in ultrasound classification) was present in 44% of infants, a stronger influence of genetic factors can be found for severe IVH (grade II–IV), occurring in approximately 60% of cases. Stage I (42% of IVH cases), defined as unilateral or bilateral reproductive matrix haemorrhage, can be excluded from association studies as a condition with low genetic impact [One limitation of this study is its relatively small sample size for genetic research into DWMI. Post hoc power analysis, however, showed a statistical power of nearly 80% for the association analyses between c impact . A signiHIF1A’s genotype and clinical risk factor profile. Since even mild hypoxia has been shown to be significantly associated with complications of prematurity, more careful monitoring of oxygen supplementation could be recommended until new effective pharmacotherapy is found. Understanding the implications of our observations is of great importance for improving the care of premature babies in intensive care units.In conclusion, our findings suggest a potential role of the hypoxia-inducible pathway in the pathogenesis of preterm complications, specifically DWMI, NEC, and RDS. Hypoxia at birth and prolonged oxygen supply were independent predictors of DWMI. Further studies with larger sample sizes are needed to confirm these findings and explore their clinical implications. Additionally, exploring the impact of these genetic variations on treatment outcomes may provide valuable insights for personalized management strategies for preterm infants. Studies focusing on the role of local or systemic pharmacological inhibition of HIF-1 and VEGF or, on the other hand, targeting the development of pharmacotherapy for HIF-1 stabilization and antioxidant supplementation, may be suggested based on a risk assessment that includes


Trimethylamine-N-oxide (TMAO), an intestinal microbiota-derived choline metabolite, has been found to be associated with ischemic stroke (IS) in more and more studies. However, the causal role of TMAO on IS occurrence remains perplexing.We comprehensively screened the related clinical studies on PubMed, Web of Science, and Embase. Case-control and cohort studies that reported the TMAO levels of both IS patients and healthy controls were included, and the risk of bias was assessed according to the criteria by the Centre for Evidence-Based Medicine in Oxford, UK. A meta-analysis of the retrieved publications was performed with a random-effect model to analyze the connection between TMAO levels and IS events. Besides, a Mendelian randomization (MR) analysis was performed to study the causal effect of TMAO on IS, with pooled data of TMAO and IS obtained from genome-wide association studies (GWAS). The following methods were used: MR-Egger, weighted median, inverse-variance weighted, simple mode, and weighted mode. The study has been registered in INPLASY (Registration number: INPLASY2023100027).P = 0.0005), while distinctive heterogeneity  was observed. Sub-group analysis revealed that the heterogeneity of the studies might be derived from the studies themselves. However, no causal effect of TMAO on IS was observed (P > 0.05) in the Mendelian randomization analysis of this study.Eight cohort or case-control studies covering 2444 cases and 1707 controls were identified. The pooled data indicated that the IS patients tended to have higher TMAO levels compared with the controls (mean difference: 1.97 μM; 95% confidence interval [CI]: 0.87, 3.07; We confirmed that IS patients tend to have higher TMAO levels than healthy individuals, while our findings of MR analysis did not support the causal role of TMAO in IS occurrence. Therefore, more studies are required for a better understanding of the relationship between TMAO levels and IS onset.The online version contains supplementary material available at 10.1186/s12883-023-03458-2. Stroke is the second leading cause of death nowadays whilst ischemic stroke accounts for 85% of stroke incidences globally . IschemiTrimethylamine-N-oxide is an organic small molecular metabolite, and is converted by gut microbiota from trimethylamine (TMA)——the waste product of TMA-containing dietary nutrients, such as phosphatidylcholine, choline, and carnitine . TMAO caIn this study, we conducted a meta-analysis based on case-control and cohort studies to investigate the association between TMAO levels and IS occurrence by pooling the current studies, and evaluate the potential of TMAO to be a biomarker that predicts IS incidences. At the same time, the causal effect of TMAO on IS was studied using a two-sample MR method.Systematic research was performed in PubMed, Web of Science, and Embase for the related studies up until September 2023. Our medical subject heading terms were “(Trimethylamine N-oxide [Title/Abstract] OR TMAO [Title/Abstract]) AND (ischemic stroke [Title/Abstract])” for PubMed, “(TS = (Trimethylamine N-oxide OR TMAO)) AND TS = (Ischemic stroke)” for Web of Science, “((Trimethylamine N-oxide or TMAO) and ischemic stroke).mp. ” for Embase. The screened references for the articles were manually checked Table S. The pubStudies were included if they were case-control, or cohort studies that reported the TMAO levels of both IS patients and healthy controls. The systematic reviews and the abstracts were excluded. Two reviewers (YC and XH) independently assessed the eligibility of the identified articles, and a third reviewer (HYR) was consulted when disagreements occurred.The inclusion criteria for the patients in this meta-analysis were: (1) studies including both consecutive patients with first-ever ischemic stroke and controls; (2) symptoms onset less than 24 h; (3) blood sample collected within 24 h post IS onset.The exclusion criteria for this study were: (1) subjects with myocardial infarction, heart failure, malignant tumor, or other systemic diseases; (2) subjects who had used antibiotics/prebiotics or experienced gastrointestinal symptoms in the past 3 months; (3) systematic reviews, meta-analysis articles, and conference abstracts.https://markummitchell.github.io/engauge-digitizer/).Two of the researchers (YC and XH) independently extracted the data including author names, year of the study, country/area, and race of the population, sample sizes of cases and controls, and characteristics of IS cases and controls. A third reviewer (HYR) was consulted when disagreements occurred. The data for TMAO levels expressed as mean ± SD were directly obtained; the data shown as median with interquartile range were converted to mean ± SD using the previously reported method ; and theThe evidence levels of identified studies were assessed according to the criteria by the Centre for Evidence-Based Medicine in Oxford, UK . The quaI2 statistic. A random-effect model was used if the studies were heterogeneous, while otherwise, a fixed-effect model was adopted. Subgroup analyses were performed to evaluate the effect of race or sample variations. We performed the sensitivity analyses by excluding one study at a time to estimate the influence of every single study on the overall estimate.All the statistical analysis was performed using Review Manager 5.3 and Stata/SE 12.0. Since the obtained data for TMAO levels were continuous, the weighted mean differences with 95% CI were used to compare the difference in TMAO between IS patients and the controls. Statistical heterogeneity among the studies was quantified using the P < 0.05 was obtained.The TMAO levels were extracted as means (SD), and the results of the meta-analysis were presented as weighted mean differences [95% CIs]. Results were considered to be statistically significant when a value of P < 5 × 10–5) were selected and clumped as instrumental variables (IVs).Summary data for TMAO were obtained from a GWAS of 217 plasma metabolites in the Framingham Heart Study including 2,076 European participants [https://gwas.mrcieu.ac.uk/) with the trait of “Ischemic stroke”, including 34,217 Europeans (id: ebi-a-GCST006908) [Summary data for IS were extracted as the outcome from the IEU GWAS database (T006908) .Inverse-variance weighted (IVW) method, the most efficient estimate in the absence of pleiotropy , was maihttps://github.com/MRCIEU/TwoSampleMR) was used to perform the statistical analysis, and the results were shown as beta ± SE. A P value less than 0.05 was considered a significant difference.The TwoSampleMR package of R (The protocol of this study has been registered in INPLASY (Registration number: INPLASY2023100027).A total of 239 records were identified from previous studies, of which 118 were duplicates to be removed. Subsequently, 63 irrelevant studies were excluded by the manual screening of the title and the abstract. Besides the mentioned databases , we have also performed a literature search in Google Scholar and got about 6,150 results, which were duplicate or irrelevant and not included here. After the full-text assessment, we finally obtained 8 publications including 4,151 cases  eligible for the meta-analysis   was observed, thus we exploited the random-effect model for the data analysis.According to the data pooled from eight studies including 4,151 cases, the IS patients tended to have higher levels of TMAO in their blood than the controls  Fig. A. To be P < 0.0001); while in the two studies in Europe, where most cases were Caucasian, no significant difference was observed  . No significant difference was exhibited in the serum sub-group    , 53 TMAO SNPs were finally retrieved. Subsequently, 52 SNPs for IS were extracted from a European-only and transancestral GWAS [The SNPs of TMAO and IS included in this study are shown in Table Sral GWAS .p = 0.315) , Maximum likelihood method , Weighted median method , and Weighted mode method .We found that TMAO did not show a statistically significant causal effect on IS, according to the MR analysis through the IVW method  or MR-Egger (p = 0.429) method (Table Sp = 0.0691) (Table SHeterogeneity was not observed through the IVW er p = 0.29 method) Table S, which i9 method er p = 0.29 methodIn the meta-analysis of eight studies involving 4,151 cases, we confirmed the positive correlation between TMAO levels and IS onset. The IS patients tend to have higher TMAO levels than the healthy people in most cases. However, the Mendelian randomization analysis in this research indicated no causal effect of TMAO on IS.Noticeable between-study heterogeneity existed in the meta-analysis, probably due to the differences in age, race, geographic location, or dietary habits of the individuals in the cohorts. Although most research indicated higher TMAO levels in IS patients, some reported controversial results: In the work by Haak et al. , the TMAThe two opposite phenomena observed in the above studies are likely influenced by a variety of factors including, but not limited to, the timing of testing, the underlying medical condition, medications being used, and dietary differences. The TMAO concentrations actually vary dynamically over time and can be affected by the time point for detection or the iatrogenic factors. It has been shown that the blood TMAO levels may be at a high level within 24 h of IS onset, and then gradually decrease , 30. TreP < 0.001), however, we cannot guarantee that our results were not affected by unmeasured confounders. There is the possibility that TMAO and IS share a similar genetic basis leading to strong clinical associations between them.It has been reported that TMAO could act as an “accomplice” in atherosclerosis, thrombosis, arterial injury, and neurological damage by promoting macrophage foam cell formation, modulating platelet hyperactivity, disrupting cholesterol metabolism, enhancing inflammation, et al. , 34. TMAThe pathogenesis of IS is quite complex and closely related to multiple risk factors, including personal characteristics like age, gender, race/ethnicity, and heredity; the underlying disease like hypertension, cardiac disease, particularly atrial fibrillation, diabetes mellitus, hypercholesterolemia; and the lifestyles such as cigarette smoking and alcohol abuse . IS occuP < 0.001) in the blood biochemistry biomarker high-density lipoprotein, and the small molecule metabolites lactic acid, pyruvic acid, and TMAO [In this study, we focused on the relationship between TMAO and IS. In fact, a large number of other small molecule metabolites may be associated with the course of IS. For example, Chen et al. compared control and mild IS patients over 3 years and found significant differences (and TMAO . Liu et and TMAO . Kijpaisand TMAO . Taken tThere are several limitations to this research. First, the lack of clinical studies limited the generalisability of our results to different ethnicities and different sample types, as only two studies of Europeans were included in the meta-analysis, and two studies used serum samples. Second, to maximize the sample size we combined studies with different study designs, and there are differences in the cohort groups, inclusion criteria, and matching strategies among these included studies, which might cause the significant between-study heterogeneity in the meta-analysis. In addition, even though we did not observe the causal association between genetically increased TMAO and IS, we cannot conclude whether the causal association exists or not, or whether TMAO and IS share a genetic basis.In summary, the meta-analysis suggests that higher TMAO levels are associated with IS occurrence, while the MR results do not support a causal effect of TMAO with IS. More studies are required to explain further insight into the relationship between TMAO levels and IS occurrence.Additional file 1: Table S1. Search stratages for literatures applied in the meta-analysis. Table S2. Genetic IVs for TMAO. Table S3. Estimation of Wald ratio between each SNP related to TMAO and risk of IS. Table S4. Heterogeneity analysis using Cochran’s Q test. Table S5. Horizontal pleiotropy analysis using the intercept of MR-Egger regression test.  Additional file 2: Fig. S1. Quality assessment of the included studies for meta-analysis. Fig. S2. Leave-one-out analysis indicated that no single SNP was driving the association of MR analysis. Fig. S3. The funnel plot for visually inspection for horizontal pleiotropy of MR analysis. Fig. S4. The Radial plots for classification of outliers in the SNPs.


Vitamin D is important for immunomodulation and may play a role in autoimmune diseases. Studies have reported a high prevalence of vitamin D deficiency in rheumatoid arthritis (RA) patients, and vitamin D status, assessed by circulating 25-hydroxyvitamin D [25(OH)D] concentration, is inversely associated with RA disease activity. However, it is unclear whether vitamin D deficiency increases the risk of later developing RA. We conducted a systematic review and meta-analysis of pre-diagnostic 25(OH)D concentrations and risk of RA.Medline and Embase databases were searched in December 2021 using various keywords for ‘vitamin D’, ‘rheumatoid arthritis’, and ‘prospective study’. Publications identified from the search were screened for eligibility, studies were excluded if vitamin D status was measured at or after RA diagnosis, and data were extracted from relevant articles. Bayesian meta-analysis was used to estimate the summary relative risk (RR) and 95% credible interval (CrI) for risk of RA in relation to circulating 25(OH)D concentrations, as well as the between-study heterogeneity.I2 = 52%). No associations were evident in men  or women .The search strategy yielded 908 records, of which 4 publications reporting on 7 studies, involving a total of 15,604 participants and 1049 incident RA cases, were included in the meta-analysis. There was no suggestion of an association between 25(OH)D concentration and subsequent risk of RA. The pooled RR per 25 nmol/L increment in 25(OH)D was 0.96 D and RA risk, but there was notable between-study heterogeneity and a lack of precision. Investigations in large-scale prospective studies with long follow-up or suitably designed Mendelian randomisation studies with consideration of potential non-linear relationships are needed to determine whether vitamin D is involved in RA aetiology.The online version contains supplementary material available at 10.1186/s41927-023-00325-y. Rheumatoid arthritis (RA) is a progressive autoimmune disease affecting around 0.5–1% of adults globally . RA maniVitamin D deficiency has been suggested to be associated with various immune-mediated inflammatory diseases including RA –13. ImmuSeveral studies have shown that vitamin D deficiency is more prevalent, and 25-hydroxyvitamin D [25(OH)D] concentrations (which provide a measure of vitamin D status) are lower in RA patients than healthy controls , 16. In An electronic literature search was performed in Medline and Embase databases to identify all eligible studies on the association between pre-diagnostic 25(OH)D concentrations and RA risk that were published from inception to 21 December 2021. The search strategy included keywords and MeSH terms covering a comprehensive list of possible variations of ‘vitamin D’ and its forms, ‘rheumatoid arthritis’, and prospective study designs. The full search strategy for Medline and Embase is listed in Additional File Results from the database searches were merged and duplicates removed using Covidence, a web-based software platform for systematic review management . RecordsData extracted from the eligible papers included: first author name, year of publication, country/location, study name or description, study period, duration of follow-up, number of participants or controls, number of cases, age and sex distribution, study design, outcome assessment, method of 25(OH)D measurement, mean/median 25(OH)D, comparison (the contrast or metric of 25(OH)D), RR/OR/HR and 95% CI, and any adjustments made for covariates. If several multivariable models were reported, the most fully adjusted effect estimate was extracted and used for the meta-analysis. If an article reported pooled results from more than one study, only the individual results were used in this meta-analysis. Data were extracted by one investigator (RC) and separately verified by two investigators (JLC and AKH).The quality of each of the studies included in the meta-analysis was assessed with the Newcastle–Ottawa scale . Each st2, the percent of total variation that is due to true between-study heterogeneity, which is derived from the posterior distribution of τ. The prior distribution for the log RR is weakly informative , based on the assumption that the RR will have a magnitude of less than 3.Prior to meta-analysis, results from all studies were converted to estimates for a 25 nmol/L increment in 25(OH)D. If a study reported results for 25(OH)D in ng/ml, concentrations were first converted to nmol/L by multiplying by 2.496. Given the low prevalence of RA, different effect measures  were considered equivalent to RRs and used directly. If a study reported results separately by sex, or other subgroups, but not overall, the subgroup-specific estimates were pooled using a fixed-effect model before inclusion in the meta-analysis. If a study reported a CI with insufficient precision (due to rounding), the CI was estimated using the reported p-value. Because few studies met the eligibility criteria for inclusion, we used a Bayesian meta-analysis approach to estimate the summary RR and 95% credible interval (CrI) for RA in relation to a 25 nmol/L increment in 25(OH)D concentration. The CrI, similar to a frequentist CI, quantifies the uncertainty around the point estimate. Heterogeneity of RRs across studies is to be expected, given the differences in populations, study designs, and methods for ascertainment of outcomes. A frequentist random-effects model can reliably handle heterogeneity in a meta-analysis of a large number of studies, however, these models are not well-suited to meta-analyses of few studies. A Bayesian approach to meta-analysis accounts for uncertainty in the estimate of heterogeneity inherent in such situations . AdditioTo investigate potential heterogeneity in the results, subgroup analyses were conducted by sex among studies that reported sex-specific estimates. To evaluate whether the results were driven by one large study or a study with an extreme result, a sensitivity analysis was performed sequentially excluding each study from the analysis one at a time and assessing the impact on the summary estimate. As a further sensitivity analysis, frequentist random-effects models were run to estimate the summary RR and 95% CI for risk of RA per 25 nmol/L increment in 25(OH)D. Risk of publication bias was evaluated by visually inspecting funnel plots and using the Egger linear regression test  and BeggAll statistical analyses were performed using R version 4.1.0 , and theThe flow diagram of study selection is shown in Fig. The characteristics of included studies are summarised in Table The average study quality score using the Newcastle–Ottawa scale was 7.7 out of 9 Table . DiffereThe seven studies included in the linear dose–response meta-analysis comprised a total of 15,604 participants and 1049 RA cases. Brink et al.  did not 2 = 52%)  or women   D was 0.96 (95% CrI 0.82–1.13) and there was moderate heterogeneity between studies D ranged from 0.94 (95% CrI 0.77–1.13) when the Nurses’ Health Study (NHS)  was omitP = 0.95) and Begg’s (P = 0.56) tests did not suggest publication bias, but the low number of studies precluded reliable assessment.Visual inspection of the funnel plot D concentration and risk of RA, but the number of studies was small, most studies contained few cases, and estimates were imprecise. Using a Bayesian approach, moderate to high between-study heterogeneity was identified.There is inconsistent evidence for a role of vitamin D in RA development. In agreement with our findings, a Mendelian randomisation analysis did not find evidence of a causal association between 25(OH)D and risk of RA . AnotherAlthough this meta-analysis did not find any association between vitamin D status and RA risk, it is possible that vitamin D plays a role in disease severity and progression. Several studies have reported favourable associations between higher 25(OH)D concentrations or vitamin D supplementation and lower disease activity in patients with RA –18. In aThis systematic review focused on the relationship between pre-diagnostic 25(OH)D concentration and subsequent risk of RA, and thus the findings cannot be directly compared with studies assessing the post-diagnostic association in patient populations. The duration of follow-up (average time between 25(OH)D measurement and diagnosis) in some of the individual studies was only a few years and it is possible that some participants might have had underlying undiagnosed disease when blood samples were collected. An inverse association between 25(OH)D and RA risk was reported among women in NHSII with 25(OH)D measured less than 4 years before symptom onset  increase in 25(OH)D), but not for 25(OH)D measurements 4 or more years prior to diagnosis . SubclinThis systematic review identified very few prospective studies that have investigated the relationship between pre-diagnostic 25(OH)D and risk of RA. The included studies had few RA cases and participants were predominantly women. Moreover, there was a lack of geographical and ethnic diversity, with results available from studies in only three countries , and almost all participants were of European descent. Further research is needed in studies with large numbers of participants that include those of different ethnicities, different geographical locations, broader age groups, and adequate numbers of cases.A limitation of this meta-analysis was the wide variability in assays used to measure 25(OH)D in the individual studies. Measurements of 25(OH)D exhibit considerable inter-assay and inter-laboratory differences, and there is a lack of standardisation of methods, which hinders the ability to directly compare results from studies that report results based on unstandardised 25(OH)D measurements , 38. We Emerging evidence suggests that associations of 25(OH)D with various health outcomes appear to be non-linear, and if vitamin D is beneficial, it is likely to be relevant only for those with vitamin D deficiency , 40. HowA strength of this meta-analysis was the use of a Bayesian approach, which allowed for a more robust estimation of the between-study heterogeneity compared with frequentist methods, which is particularly useful when there are few studies included.In summary, this systematic review and meta-analysis found no association between vitamin D status and risk of RA. Nevertheless, few studies have investigated this relationship, and no firm conclusions can be drawn in the context of limited data. Further large-scale prospective studies with long follow-up durations, a larger number of cases, and enriched with participants with a broad range of 25(OH)D (particularly those with very low concentrations) are warranted to determine whether vitamin D deficiency is associated with the risk of developing RA.Additional file 1. Medline search strategy. Embase search strategy. Figure S1. Frequentist random-effects meta-analysis of 25-hydroxyvitamin D concentration and risk of rheumatoid arthritis. Figure S2. Linear dose-response meta-analysis of 25-hydroxyvitamin D concentration and risk of rheumatoid arthritis, omitting each individual study one at a time. Figure S3. Funnel plot for the meta-analysis of 25-hydroxyvitamin D concentration and risk of rheumatoid arthritis.


By 2030, the World Health Organization wants to decrease viral hepatitis incidence and mortality by 90% and 65%, respectively. One of the agents responsible for the increased burden of viral hepatitis is the hepatitis E virus (HEV). This emerging pathogen is prevalent worldwide causing both acute and chronic infection. The rising risk profile of HEV has become a source of increased global public health concern. Despite this challenge, South-Eastern Asia (SEA), where many at-risk people are found, lacks uniform HEV prevalence data. Therefore, a meta-analysis was conducted to assess the overall prevalence of hepatitis E in SEA. Using R statistical software, a random effect model was used to estimate the logit-transformed prevalence. Moderator analyses were used to investigate the potential sources of variation. Thirty-two studies comprising 29,944 with 6806 anti-HEV antibody-positive individuals were evaluated. The overall HEV seroprevalence in SEA was 21%  with high heterogeneity. At the country level, Laos has the highest prevalence estimate of 39% (CI: 16–69). Also, the studied population, year of publication, duration of sampling, and diagnostic method are significant HEV prevalence predictors accounting for 22.61% of the observed heterogeneity. The high HEV prevalence found in this study necessitates coordinated national and regional efforts to combat this emerging disease. Some of these agents cause infections that may result in severe complications such as hepatocellular carcinoma or even death2. According to estimates from the World Health Organisation (WHO), viral hepatitis is responsible for the death of approximately 1.45 million people annually2. About 5–10% of those fatalities are attributed to hepatitis A and E, while the remaining are caused by hepatitis B and C2. Thus, hepatitis E virus (HEV) is now considered a significant contributor to the rising global burden of viral hepatitis.Globally, millions of people are infected with viral hepatitis yearly; a significant public health concern. The primary known aetiological agents of viral hepatitis are hepatitis A, B, C, D, and E viruses3. Five of these genotypes (HEV-1 to HEV-4 and HEV-7) are known to be of human clinical importance. According to seroprevalence studies, a third of the world’s population is at risk of HEV infection3. Equally, annual new infections, including symptomatic cases, are increasing worldwide. Some infections also result in fatalities, with HEV accounting for 3.3% of all viral hepatitis mortalities globally3. Higher seroprevalence rates are recorded in developing countries than the developed nations4. However, epidemics are restricted to only developing regions, particularly in Africa and Asia4. The increasing global hepatitis E burden can be attributed to the multiple transmission routes of HEV. These transmission routes include water, food, and blood-borne as well as vertical and zoonotic4. Although most infections by HEV are usually mild and self-limiting4, certain groups are at risk of severe and or chronic disease. Acute HEV infection in pregnant women, mostly in developing countries, may present with fulminant hepatic failure with high mortality rates5. At the same time, chronic disease is seen in immunocompromised populations, mainly in developed regions4. Thus, HEV seriously threatens global health in developed and developing nations.Furthermore, HEV genotypes (HEV-1 to HEV-8) are being identified from diverse hosts1, providing a comprehensive information on the epidemiology of hepatitis E is critical. The need for this critical data is even more pertinent in the South-Eastern Asia (SEA) sub-region. This objective can be achieved efficiently through conducting a systematic review and meta-analysis. Thus, this study was undertaken as a follow-up to the previous systematic review (SR)6 by pooling the estimated effect of the seroprevalence of HEV from the included studies. Also, to investigate possible sources of heterogeneity through moderator analysis.So, with the renewed effort by WHO to reduce the global burden of viral hepatitis6. The earlier review was conducted according to the PRISMA  was obtained by adopting the random effect (RM) model. The generalised linear mixed-effect model (GLMM) method11 was used to pool the effect estimates. The confidence interval around the pooled effect was calculated using the Knapp–Hartung adjustment.The R studio desktop programme (version 2020.02.3 + 492) through the R software (version 4.1.2-2021-11-01) environmentX2, Cochrane Q test, I2, and T2 statistics using restriction maximum likelihood (REML) estimation to calculate heterogeneity variance. The Cochrane Handbook of Systematic Reviews of Intervention12 guideline served as a reference for interpreting the heterogeneity threshold. Thus, an I2 value of 0–40% was considered not important; 30–60% represented moderate heterogeneity; 50–90% was regarded as substantial heterogeneity; 75–100% was deemed considerable heterogeneity. While a p-value of < 0.05 was interpreted as indicating significant heterogeneity. A 95% prediction interval (PI) was also estimated to determine the effect size variation across studies13.Heterogeneity was estimated by 14 through the “find.outlier” function of “dmetar” package of R (version 4.1.2-2021-11-01) software. Influential studies that may push the effect estimate into one direction were investigated by conducting influence diagnostics using the “influential analyses” function of “dmetar” and “metafor” packages of R (version 4.1.2-2021-11-01) software. The influence diagnostics include the Baujat plot15, Viechtbauer–Cheung influence plot14, graphic display heterogeneity (GOSH) plot16, and leave-one-out meta-analysis.Outlier and influence analyses were conducted to determine the robustness of pooled estimate and assess each study's influence on the summary proportion. Studies with extreme effect sizes were identified as outliers17. The variables used for the analyses include country of study, location setting , diagnostic method, assay type, and studied population. Others are gender, age, year of publication, sample size, and duration of sampling (in years). The age variable was determined for each of the included studies by the reported mean age. While gender was captured based on the predominant gender of the study population as reported in the study. The studied population variable was categorised as healthy, clinical, or mixed depending on whether a healthy general population, hospital-based (patients) or both respectively were sampled. For meta-regression, each of the covariates was assessed in univariate analysis. Only predictors with p < 0.25 were included in the multivariate meta-regression analysis model.Subgroup and meta-regression analyses were conducted to investigate the possible sources of the observed heterogeneity in the meta-analysis. Variables considered for evaluation as potential moderators were based on the reports from previous studies18. Egger’s regression test was further used to investigate the significance of the observed asymmetry19.Potential publication bias was assessed by constructing and observing the funnel plot of the logit-transformed prevalence against standard error6. Thirty-two studies from seven SEA states are included in the meta-analysis out of the 35 studies screened for eligibility  study  clustering: Study 25, Study 1, Study 8, Study 10, Study 18, Study 23, Study 19, and Study 2849.Gaussian Mixture Model (GMM) clustering: Study 25, Study 1, Study 8, Study 10, Study 18, Study 23, Study 19, and Study 28The plot revealed distinct effect size–heterogeneity clustering patterns of (1) high effect size–high heterogeneity  of heterogeneity.2 = 0.2677; tau = 0.5174; I2 = 97.6%. Other countries' seroprevalences and their respective statistics are presented in the subgroup analysis forest plot  had the highest seroprevalence of 39%  local production system61. However, studies that combined urban and rural populations produced the highest prevalence. This phenomenon can also be explained by the increase in rural–urban migration and the possibility of sampling high-density urban areas61.Living in rural areas has been established in numerous studies as a risk factor for HEV infection. Thus, expectedly, higher hepatitis E prevalence were discovered in the rural than the urban areas. Other associated factors might have contributed to the observed outcome besides the earlier enumerated risk factors. The populace may have lower educational and economic status in rural areas than the urban settlers62. Thus, assays have predictive and modifier effects on HEV seroprevalence17. Therefore, the findings of this meta-analysis agree with several primary studies and meta-analyses63. The result did not show any specific pattern in sample sizes, probably due to the relatively small number of studies used in most groupings.Similarly, this study shows that reported seroprevalence rates were impacted by the employed diagnostic method and assay type. Studies using total HEV antibodies as diagnostic methods showed higher prevalence than those using only IgG. Also, there was observed variation in HEV prevalence among the evaluated assays. The performance of assay types is often influenced by their specificity and sensitivity levels. HEV assay performance comparative studies have shown varied seroprevalence rates among the assays evaluated4. The most suggested reason for the propensity is the increased behavioural exposure in men compared to women4. This study also revealed a pattern that suggests an association of anti-HEV antibody positivity with increasing age. Prevalence was lowest in the age group of 20–25 years and peaked at 44–49 years age group. The observed pattern conforms with many primary and secondary studies52. Some explanations for this pattern include lifetime cumulative exposure and the impact of ageing on immunological function52. Others are genomic instability and other ageing processes; thus, prevalence tends to increase with age52.Furthermore, this current study also showed that the HEV seropositivity rate is lower in the healthy population compared to the clinical and mixed populations. The result further established that those with underlying diseases and the immunocompromised have an increased risk of HEV infection compared to the healthy population. Differences were also observed between genders, ages, years of publication, and sampling duration. The prevalence was higher in males than females, as previously established. Studies have reported higher HEV seroprevalence in males than non-pregnant females64. This phenomenon may result in more human infection due to frequent contact with these animal hosts of the pathogen. Also, in recent years, there seems to be an increased awareness among researchers about HEV being a critical causative agent of viral hepatitis.Evaluation based on the year of publication showed almost a two-fold increase in HEV seroprevalence rate in 2019 from the 1999 rate. Likewise, it was observed that the longer the duration of sampling, the higher the prevalence. Studies with longer sampling duration had older samples indicating that studies with more recent samples had lower prevalence. This increase could be attributed to a substantial rise in HEV infection, increased awareness and research, or the use of more sensitive diagnostic assays. Although there are concerns that hepatitis E is still underestimated, recent discoveries about the epidemiology of the disease might have contributed to the observed rise in its prevalence as detected. These discoveries have led to a series of epidemiologic and pathological studies aimed at understanding the nature of the disease. For example, studies have shown constant expansion of HEV’s host range with an increased probability of cross-species infections65. The use of more effective diagnostic assays will be more accurate compared to the less sensitive techniques. These factors, individually or in combination with the contribution of others, might have been responsible for the observed rate rise reported in this study. However, with increased awareness, clinical evaluation, and research, the prevalence might increase, as indicated by the prediction interval estimate.Additionally, most recent assay types have shown improved efficacy in diagnosing the disease over the oldOn a more technical note, it was observed that substantial variation persisted in each of the assessed groups after subgroup analyses. Although heterogeneity is expected in the pooled effect size, it is assumed that the variation should be reduced significantly in close groups. Thus, its persistence in homogeneous subgroups calls for deeper exploration. So, further assessment using meta-regression analysis showed four factors  that are significant moderators of the effect size. The quantifiable factors are studied population, diagnostic method, year of publication, and sampling duration.66. However, the assessment showed that publication bias was not present in this survey. Though, as already established, heterogeneity could be due to quantifiable, hidden, or random moderators66. Other hidden factors that were not evaluated such as HEV genotype variation, might be responsible for the residual heterogeneity. Some others might even be entirely unknowable thus, cannot be specified or controlled66. So, pooled studies in meta-analyses can vary due to reasons that will likely never be fully identified66.Nevertheless, the predictors could only account for a fraction of the variation. Other factors that can potentially bias effect size estimates, such as file drawer effect, p-hacking strategies and publication bias, may equally affect the heterogeneity estimate67.Consequently, primary prevalence studies must be designed as similarly as feasible to reduce between-study variation. The deduced recommendation from the findings of this study is for future seroprevalence studies to sample homogenous populations. Subsequent surveys at either national or regional levels can adopt a unified diagnostic method for hepatitis E investigation. In addition, as recommended by different researchers, further primary studies should consider regional and national surveys using similar study designs and analytical methodsAs the first meta-analysis of hepatitis E seroprevalence in SEA, this study has pooled many studies, leading to a relatively large sample size with enhanced statistical power. Also, this experiment was conducted using meticulous methods, and moderators were thoroughly investigated. Therefore, a comprehensive sub-regional as well as national data on the seroprevalence of HEV is provided. The provided information will thus assist further research and informed decision-making for designing HEV preventive and control measures in the sub-region. However, there are a few limitations of this study. Included studies are only from seven countries due to the non-availability of studies from the remaining four countries. This constraint may affect the generalisation of the result to represent the sub-region. Likewise, fewer studies were available in some of the groups during subgroup analyses which may affect the estimates in the respective groups. Thirdly, only HEV seroprevalence was considered in this analysis; sporadic HEV infection and epidemics that equally contributes to the disease burden were not estimated.Even though there are no reports from some countries in the sub-region, SEA has high HEV seroprevalence. As expected, the prevalence is higher in some countries than others, and the variation is attributable to detectable, concealed, and random factors. Therefore, there is a need for concerted efforts towards preventing and controlling this emerging disease at national and regional levels. Increased research, surveillance, and purposeful screening of at-risk groups and blood donors will assist in prevention and control.Supplementary Information 1.Supplementary Information 2.Supplementary Information 3.Supplementary Figure 1.Supplementary Figure 2.Supplementary Figure 3.Supplementary Figure 4.


Polycystic ovary syndrome (PCOS) is a prevalent endocrine disorder in females. Nesfatin-1 is a neuropeptide synthesized in the hypothalamus and other peripheral organs, and there are conflicting opinions about its correlation with PCOS.This study aims to investigate the correlation between nesfatin-1 and PCOS and evaluates the effectiveness of nesfatin-1 as a biomarker for the detection of PCOS in women. A systematic review and meta-analysis were conducted to identify pertinent articles from databases such as PubMed, Web of Science, Cochrane, EMBASE, Scopus, and Google Scholar. The standardized mean difference (SMD) and 95% confidence interval (CI) were calculated using a random effects model to compare group outcomes. Additionally, meta-regression and subgroup analysis were performed to elucidate sources of heterogeneity.The meta-analysis involved 12 studies with 1222 participants, and the findings revealed a significant relationship between PCOS and nesfatin-1 levels. The pooled  indicated a significant difference between the evaluated groups. Moreover, a subgroup analysis showed that there was a substantial difference in nesfatin-1 levels among women with PCOS and higher homeostatic model assessment for insulin resistance ratio (SMD = 1.46; 95% CI: 0.92-2.00; p Our meta-analysis indicates an association between high nesfatin-1 levels and PCOS. This suggests a potential role of nesfatin-1 in the development of PCOS and proposes it as a potential diagnostic biomarker for the disease. However, further research is necessary to validate these findings. Polycystic ovary syndrome (PCOS) is a frequently encountered endocrine disorder affecting women during their reproductive years (1). The prevalence of PCOS alters among different races and countries depending on environmental and genetic factors. This syndrome is distinguished by irregular ovulation, hyperandrogenemia, and reduced fertility. It is often associated with metabolic conditions such as hyperinsulinism, obesity, and insulin resistance (IR) (2). Obesity is a critical concern in women diagnosed with PCOS, as it constitutes the most common clinical problem affecting a significant proportion of patients . Furthermore, IR is an important index of interest in research studies concerning women with PCOS.Research indicates that high insulin levels are directly associated with the severity of the disease . However, the pathophysiology of PCOS remains complex and unclear, regulated by various hormones, and possibly influenced by genetic and environmental factors . Peptides responsible for the feeling of satiety in the hypothalamus have demonstrated their effectiveness in regulating energy balance, glucose metabolism, and obesity.In addition, they have a significant impact on gonadal function, playing a crucial regulatory role in both physiological and pathological processes in the ovaries (9). Nesfatin-1 is an example of an anorexigenic peptide that could be linked to the pathophysiology of PCOS . Nesfatin-1 derives from the nucleobindin-2 gene and contains 82 amino acids.The primary source of this hormone is the central nervous system, especially the hypothalamus. Nesfatin-1 is also expressed in peripheral systems such as adipocytes, gastro-endocrine cells, and pancreatic beta cells . It has been identified as having a role in regulating feeding behavior and energy metabolism (14). Furthermore, it has been observed to have an anorexic effect by decreasing the frequency of meals and increasing the time between meals (15). Numerous studies have revealed that nesfatin-1 is closely linked to body mass index (BMI), glucose, insulin metabolism, and IR . The significant correlation between obesity and IR with PCOS has prompted extensive research into the involvement of nesfatin-1 in energy metabolism, insulin secretion, and food intake .Interestingly, different studies have reported varied findings regarding the levels of nesfatin-1 in women with PCOS, while some studies indicate higher levels ; others suggest lower levels (22).This study aims to perform a systematic review and meta-analysis to analyze the variations in nesfatin-1 levels between individuals diagnosed with PCOS and healthy controls. The main objective is to accurately assess the association between nesfatin-1 and PCOS disease.In this systematic review and meta-analysis, we conducted a literature search to identify articles related to the research topic. This review was conducted based on PRISMA statement (23). We searched the PubMed, Web of Science, Scopus, Cochrane, EMBASE, and Google Scholar databases for articles published in English from January 2000-April 2023. The search terms included PCOS  and “Nesfatin-1” or “Nesfatin-1 protein” or “NEFA protein” or “NUCB2 protein” or “nucleobindin 2 protein”, and the search was limited to humans. We preselected all articles that were potentially related to PCOS.The relevant observational studies were selected and reviewed manually and independently by 2 principal investigators, without relying on any automation tools. In addition, the researchers thoroughly reviewed the reference lists of identified articles to ensure that all relevant studies were included. In instances where discrepancies arose, they were resolved through discussions or consultations with a third supervising researcher to ensure the accuracy and quality of the study selection process.The inclusion criteria for the studies analyzed in this research were as follows: “(i) publication in the English language, (ii) diagnosis of PCOS in women of reproductive age according to the Rotterdam criteria, and (iii) measurement of nesfatin-1 as one of the biomarkers”.The following exclusion criteria were applied: (i) studies with inaccessible full texts, (ii) studies that did not report the level of nesfatin-1, (iii) studies with insufficient data, (iv) conference proceedings, reviews, randomized controlled trials, and animal studies, (v) studies without control groups.The following data were extracted according to a pre-prepared checklist: origin of study characteristics , study design , participant characteristics , measurements, diagnostic criteria, and reported outcomes. These details are shown in table I.The quality of the final studies was assessed using the Newcastle-Ottawa scale, which is commonly used to assess nonrandomized studies in meta-analyses. The article consists of 8 sections categorized into 3 main areas: selection (comprising 4 points), comparison (comprising 2 points), and exposure (comprising 3 points) (24). The scoring scale ranged from 0-9, where studies scoring 22 statistics were used to assess heterogeneity between studies, and sources of heterogeneity were identified by subgroup analysis and meta-regression. Sensitivity analysis was conducted by individually removing each study to evaluate their impact on the overall results. Statistical analysis was performed using comprehensive meta-analysis version 3 software, developed by Biostat Inc., USA, in 2018. The presence of publication bias was assessed using the Begg's and Egger's tests. The significance level was set at p The relationship between serum nesfatin-1 level and women with PCOS was assessed using standardized mean difference (SMD) with 95% confidence intervals (CI). The statistical significance of heterogeneity was defined by considering p The search strategy employed retrieved 311 articles from the database; 93 duplicates were eliminated before the screening stage. Further screening of titles and abstracts led to the exclusion of 121 articles. The remaining 97 potentially relevant articles underwent a thorough full-text review. Of these, 56 studies were excluded because of a discrepancy between their study design and the comparison group, 10 records were excluded due to discrepancies in input languages other than English, and other records were omitted because crucial information was not available. Therefore, only 12 studies with 1222 participants met our selection criteria, as shown in figure 1.A total of 12 studies were included in the analysis, which reported changes in nesfatin-1 levels in both PCOS groups and corresponding control groups.The strength of the association was measured by calculating the SMD. The result showed a higher nesfatin-1 level in the PCOS group compared with the control group . In addition, a significant heterogeneity was observed between studies , our estimation suggests that the PI ranges from -1.59-2.67. This interval implies that the true effect size in 95% of all comparable populations would lie within this range. In other words, there is a high possibility of the actual effect size in each population falling within this interval. It is important to note that this PI provides a range of plausible values for the true effect size rather than a point estimate. This means the true effect size may vary within this interval based on factors such as sample size, study design, and population characteristics. Hence, caution must be exercised while interpreting the findings within this PI .Subgroup analyses were conducted for each study based on BMI, age, and homeostatic model assessment for IR (HOMA-IR). Within the age subgroup analysis, significant differences were found between women older than 25 yr  and women younger than 25 yr , as shown in figure 4.2 or 2. Women with a BMI 2 exhibited significantly higher nesfatin-1 levels , as illustrated in figure 5.Subgroup analysis was performed based on studies with mean BMI either A subgroup analysis was conducted based on studies with a mean HOMA-IR of either Meta-regression analysis revealed no considerable association between nesfatin-1 levels and the year of study , BMI , or mean age . Nevertheless, HOMA-IR was significantly correlated with TSH  .Begg's and Egger's test were utilized to evaluate publication bias in the studies included in the meta-analysis. Our findings did not demonstrate the presence of publication bias in the meta-analysis for SMD of nesfatin-1 level  . We conducted sensitivity analysis to ensure the dependability and consistency of our results, the overall effect size remained unchanged after eliminating individual studies, indicating that our analysis was dependable.PCOS is a prevalent reproductive condition with substantial health, psychological, and economic implications for women and society . Furthermore, various markers have been introduced to assess this condition . The scattered and conflicting reports regarding nesfatin-1 levels in women with PCOS have led to extensive research investigating nesfatin-1 levels (9). Several studies have indicated significantly higher nesfatin-1 levels in PCOS patients than women with no PCOS . Other studies, in contrast, have found no significant differences in nesfatin-1 levels between PCOS and non-PCOS patients . This meta-analysis is the first study to examine the association between nesfatin-1 levels and PCOS. Our findings and analysis revealed that nesfatin-1 levels were elevated in PCOS patients compared to individuals without PCOS. Since nesfatin-1 is derived from the adipose tissue and hypothalamus within the body , and considering the association of PCOS with obesity (3), it is plausible that the correlation between nesfatin-1 and PCOS can be elucidated through this endocrine pathway. However, more research is necessary to comprehensively understand the cellular and endocrine mechanisms implicated in this metabolic pathway. Subgroup analysis revealed a significant increase in nesfatin-1 levels in individuals aged 25 yr or older compared with individuals younger than 25 yr. In addition, studies have consistently shown that pregnant women with PCOS and advanced maternal age have higher nesfatin-1 levels, potentially affecting maternal and fetal health. These findings suggest that the combination of PCOS and older maternal age can influence nesfatin-1 production during pregnancy, potentially affecting appetite and metabolic processes in both the mother and the child (35). The implications of these findings for maternal and fetal health should be carefully considered. Similarly, individuals with a BMI We performed subgroup analysis and employed meta-regression techniques to determine the origin of heterogeneity. When a factor demonstrates a significant correlation with heterogeneity, this allows for a more accurate assessment of effect size and supports the conclusion. In particular, a meta-regression analysis was conducted for the nesfatin-1 level using research year, BMI, age, and HOMA-IR as variables. Our findings revealed that age, BMI, and research year did not correlate with nesfatin-1 levels. Still, HOMA-IR was significantly associated with nesfatin-1 levels, and its increase considerably amplified nesfatin-1 levels, ultimately leading to the heterogeneity of the studies.There are some limitations to the current research, including the fact that the results are confined to a specific population and geographic location. Moreover, this study highlights the importance of examining the correlation between nesfatin-1 levels and the timing of blood sampling in relation to the menstrual cycle. As a result, it was difficult to analyze this relationship due to limited availability of data and potential bias in the chosen studies. In addition, non-English articles or dissertations were not incorporated into the study. Despite these limitations, this research provides valuable insight into the relationship between nesfatin-1 levels and PCOS. Future studies may extend these findings by investigating the endocrine circuits and cellular mechanisms underlying this disorder.This meta-analysis is the first attempt to investigate the potential association between nesfatin-1 levels and PCOS. Our research findings indicate that individuals with PCOS exhibit elevated levels of nesfatin-1 compared to those without PCOS. This is particularly evident among individuals with higher HOMA-IR index and IR. Despite the potential limitations of the study, these findings provide crucial insights that could assist in the development of novel treatments and management strategies for PCOS in the future.The authors declare that there is no conflict of interest.


Diabetes mellitus has been associated with increased breast cancer (BC) risk; however, the magnitude of this effect is uncertain. This study focused on BC risk for women with type 2 diabetes mellitus (T2DM).Two separate meta‐analyses were conducted (1) to estimate the relative risk (RR) of BC for women with T2DM and (2) to evaluate the risk of BC for women with T2DM associated with the use of metformin, a common diabetes treatment. In addition, subgroup analyses adjusting for obesity as measured by body mass index (BMI) and menopausal status were also performed. Studies were identified via PubMed/Scopus database and manual search through April 2021.A total of 30 and 15 studies were included in the first and second meta‐analyses, respectively. The summary RR of BC for women with T2DM was 1.15 . The subgroup analyses adjusting BMI and adjusting BMI and menopause resulted in a summary RR of 1.22  and 1.20 , respectively. For women with T2DM, the summary RR of BC was 0.82  for metformin users compared with nonmetformin users.Women with T2DM were more likely to be diagnosed with BC and this association was strengthened by adjusting for BMI and menopausal status. No statistically significant reduction of BC risk was observed among metformin users.These two meta‐analyses can inform decision‐making for women with type 2 diabetes regarding their use of metformin and the use of screening mammography for early detection of breast cancer. Relative risk of breast cancer among women with type 2 diabetes . For instance, type 2 diabetes is a chronic inflammatory disorder and is associated with inflammatory cell infiltrations, commonly seen in adipose tissue, which might lead to an increase in aromatase expression and increased local estrogen production. Increased estrogen production may drive the growth of estrogen receptor‐positive (ER+) breast cancer.Three peer‐reviewed meta‐analysis studies, all published prior to 2012, examined the relationship between breast cancer risk and diabetes . All three studies found that women with diabetes had an increased risk of breast cancer .The objective of the first part of the present study is to estimate the risk of breast cancer only for women with type 2 diabetes. The objective of the second meta‐analysis in the present study is to estimate the breast cancer risk for women with type 2 diabetes associated with metformin use. Subgroup analyses with a focus on menopausal status and obesity had been planned depending on the availability of data.The objective of the second meta‐analysis in the present study is to estimate the breast cancer risk for women with type 2 diabetes associated with metformin use. Subgroup analyses with a focus on menopausal status and obesity had been planned depending on the availability of data. Metformin is commonly prescribed to adults with type 2 diabetes, and around 40% of adults with type 2 diabetes in the USA take metformin as a treatment.2Following the guidance in the Cochrane Handbook,2.1We searched the PubMed and Scopus databases through April 2021 for studies published in English using the search terms summarized in Table 2.2≥30 if the type of diabetes was not specified). We used 30 as an age cutoff which is a common age cutoff to distinguish type 2 from type 1 diabetes in previous studies.The meta‐analysis for the RR for breast cancer for women with type 2 diabetes compared with women without diabetes included (1) any observational study reporting an odds ratio (OR), RR, or hazard ratio (HR) estimate with a 95% confidence interval (CI), and (2) for females diagnosed with type 2 diabetes (or diagnosed at age ≥30 if the type of diabetes was not specified).The meta‐analysis for the RR of breast cancer for women with type 2 diabetes associated with metformin use included (1) any observational study reporting an OR, RR, or HR estimate with 95% CI, (2) comparing metformin users to average nonmetformin users, and (3) included studies of females diagnosed with type 2 diabetes (or diagnosed at age Two independent investigators (Y.L. and A.H.) identified eligible studies by screening titles and abstracts. Disagreements related to the selection of the studies were resolved by a consensus decision after discussion.2.3Two investigators (Y.L. and A.H.) independently reviewed the selected studies and extracted the relevant information including full citation , geographic location of participant recruitment, age of the study population, adjustment of covariates, risk estimate, and the corresponding 95% CI. When one study reported multiple HRs, RRs, or ORs, the estimate adjusted for the highest number of covariates (first priority) for the longest duration of follow‐up was used (second priority). When multiple publications used the same cohort, only the result from the most recent publication was included, and if the previous publications reported the results by adjusting for different sets of covariates, they were used in the subgroup analysis.Two investigators (Y.L. and A.H.) independently evaluated the study quality using the Newcastle‐Ottawa Assessment scale, an assessment for observational studies, in terms of selection, comparability, and exposure for cohort studies and selection, comparability, and outcomes for case–control studies.2.4I2 test was used to evaluate heterogeneity across studies, where the recommended categories are set as possibly not important for I2 = 0%–40%; moderate for I2 = 30%–60%; substantial for I2 = 50%–90%; and considerable for I2 = 70%–100%.p value of <0.05.All extracted data were analyzed by Revman5 and MedCalc.33.1Of 3,161 possible studies identified through database search, 34 relevant studies of the effect of type 2 diabetes on breast cancer were found after screening abstracts and titles, and any nonobservational study was excluded during this process. Two additional studiesThe quality of the included studies was moderate with an average score of 6 using the Newcastle‐Ottawa Assessment scale. Of the 30 studies, 11 studies had high quality (scoring between 7 and 9); 16 studies had moderate quality (scoring between 5 and 6); and three studies had low quality (scoring below 4) . The summary risk estimate using only the cohort studies  was consistent with that using only the case–control studies  Figure .A limited number of studies presented results according to race or ethnicity, and subgroup analyses were conducted using the 20 studies in which women were randomly sampled in European and North American countries. A slightly greater RR was observed among the seven studies that provided estimates adjusted for BMI . For the three studies adjusting for BMI and menopausal status, the summary RR was 1.20 . Furthermore, some studies reported their estimates stratified by menopausal status. The summary RR was 1.07  and 0.97  for the risk of breast cancer associated with type 2 diabetes for post‐ and premenopausal women with diabetes, respectively Table .3.2For the second meta‐analysis, we identified 724 studies to estimate the risk of breast cancer according to metformin use for women with type 2 diabetes through a database search, and 17 studies were selected after screening abstracts and titles. One studyI2 was 97%. Therefore, the random effects model was used in the analysis. A funnel plot was produced, and no obvious evidence of publication bias was observed as the p value was 0.65 from Egger's test  Figure . The lim4Our analysis found that women with type 2 diabetes were more likely to have breast cancer . A stronger association was observed after the adjustment of BMI or BMI and menopausal status . For women with type 2 diabetes, our meta‐analysis suggested that metformin use was associated with a reduced risk of breast cancer although the variation around the risk estimate was large .We found three meta‐analysis studies on the relationship between breast cancer risk and diabetes published since 2010. Note that while none of the three previous reviews considered a quality assessment for the included studies, we assigned a score to each study using Newcastle‐Ottawa Assessment scale to represent its quality. This approach has led to a more reliable estimation of the RR. More specifically, three studies in our meta‐analysis were evaluated as low quality,The study published by Boyle et al.The overall estimates on summary RR from the three meta‐analyses are higher than our summary RR estimate as we limited the study population to female patients with only type 2 diabetes. A subgroup analysis on women with only type 2 diabetes by Boyle et al.Boyle et al.Obesity is a confounding factor for the association between diabetes and the risk of breast cancer.There are some limitations to our analysis. In addition to obesity, menopausal status is another factor that might affect the final estimate.We also did not account for the differences in mammography screening utilization between women with and without diabetes. We do not expect the control on mammography screening utilization would affect our results. This is because although women with diabetes have higher healthcare utilization than women without diabetes, the complexity of diabetes care often decreases the rates of mammography screening for women with diabetes.Two review studies examined the risk of breast cancer for women with type 2 diabetes associated with metformin use. The review by Tang et al.Another study of the association between metformin use and breast cancer risk by Yang et al. included 11 estimates from nine studies, which reported a summary RR of 0.96 . Among the nine studies, Tsilidis et al.,Our meta‐analysis included four additional studies published after 2016. However, similar to these studies, while we found a slightly lower summary RR , the association was not statistically significant.The summary estimate was highly driven by the estimate from Tseng et al.,p value from Egger's test was 0.65. Although two prior reviews included estimates compared with other glucose‐lowering medicines (sulfonylurea or insulin‐based treatments), we only selected studies having nonmetformin users as the reference group. Furthermore, although Egger's test can be used to assess the asymmetry of funnel plots, it works better for continuous outcomes with intervention effects measured as mean differences.Unlike the previous reviews, no obvious evidence of publication bias among the studies was observed as the Overall, obesity and menopausal status could serve as confounding factors that may have affected our results. However, a limited number of observational studies adjusted for these two factors, and thus, future work could examine the effects of these two factors. Also, based on the estimates from these two meta‐analyses, another future research direction could be to optimize the decision‐making for women with type 2 diabetes regarding their use of metformin and the use of screening mammography for early detection of breast cancer.5Overall, we found women with type 2 diabetes were about 15% more likely to be diagnosed with breast cancer than women without type 2 diabetes. A stronger association (RR = 1.22) was observed after the adjustment of BMI and menopausal status. Metformin use was not associated with a statistically significant reduction in breast cancer risk.Yifan Lu: Conceptualization ; formal analysis (lead); investigation (lead); methodology (lead); writing – original draft (lead). Ali Hajjar: Formal analysis (supporting); investigation ; validation (lead). Vincent Cryns: Resources ; writing – review and editing . Amy Trentham‐Dietz: Funding acquisition ; resources ; writing – review and editing . Ronald E. Gangnon: Resources ; writing – review and editing . Brandy Heckman‐Stoddard: Writing – review and editing (supporting). Oguzhan Alagoz: Conceptualization ; funding acquisition ; project administration (lead); supervision (lead); writing – original draft .This work was supported by National Cancer Institute grants U01 CA253911 and P30 CA014520.OA has been a paid consultant for Bristol Myers Squibb, Johnson & Johnson, and Exact Sciences, outside of the submitted work. No other conflicts to report.Figure S1.Click here for additional data file.