P=0.017 for GG compared with AA+GA; OR = 1.81, 95% CI = 1.17–2.82, P=0.008 for GG compared with GA). The GG genotype of the ESR1 XbaI polymorphism could be a genetic risk factor for severe pre-eclampsia susceptibility. However, the ESR1 gene PvuII -397T/C polymorphism was not significantly associated with the risk of severe pre-eclampsia, and there was no association between mild pre-eclampsia and the ESR1 gene PvuII -397T/C and XbaI -351A/G polymorphisms separately. The current meta-analysis indicates that the ESR1 XbaI genetic polymorphism may be associated with severe pre-eclampsia. However, there was no association of the ESR1 gene PvuII and XbaI polymorphisms with the risk of mild pre-eclampsia. Owing to the low statistical power, the results may not be sufficiently robust and this conclusion should be interpreted cautiously, which highlights the requirement for large-scale and high-quality studies in this field.This meta-analysis was performed in order to determine the associations between the estrogen receptor α (ESR1) gene PvuII site  and XbaI site  polymorphisms with severe and mild pre-eclampsia. Eligible studies were identified by searching PubMed, Medline, Embase, China National Knowledge Infrastructure (CNKI), and WanFang databases until May 2018. The pooled odds ratio (OR) and 95% confidence interval (CI) were used to calculate the associations. Six articles  investigated the ESR1 gene PvuII -397T/C and XbaI -351A/G polymorphisms in severe and mild pre-eclampsia patients and included controls. The pooled results indicated an increased risk of severe pre-eclampsia for the XbaI -351A/G polymorphism (OR = 1.67, 95% CI = 1.10–2.25,  Pre-eclampsia is a major leading cause of maternal mortality worldwide 2–5%), affecting women after 20 weeks of pregnancy, and is characterized by increased systemic vascular resistance, decreased blood volume, vascular endothelial cell destruction, and renal hemodynamic abnormalities . Annuall–5%, affeThe estrogen receptor α (ESR1) gene is located on the long arm of chromosome 6 (6q25.1) and contains eight exons . ESR1 isThe association between ESR1 PvuII -397T/C (rs2234693) and XbaI -351A/G (rs9340799) polymorphisms with the risk of severe and mild pre-eclampsia was found to be inconsistent. Therefore, a meta-analysis of six articles (seven studies) was performed to investigate the association between ESR1 polymorphisms and the risk of severe and mild pre-eclampsia.Literature published until May 2018 was retrieved from the PubMed, Medline, Embase, China National Knowledge Infrastructure (CNKI), and WanFang databases, both English and Chinese language articles were included. The search strategies contained combinations of the following terms: , (‘polymorphism’ or ‘variant’ or ‘mutation’), and (‘pre-eclampsia’ or ‘severe pre-eclampsia’ or ‘mild pre-eclampsia’). Reference lists were also retrieved and manually searched to identify additional potential studies.Included studies met the following criteria: (i) assessing the association of ESR1 -397T/C PvuII (rs2234693) or -351A/G XbaI (rs9340799) polymorphisms with the risk of pre-eclampsia, (ii) sufficient data to calculate the odds ratio (OR) with 95% confidence interval (CI), and (iii) case–control study.Studies were excluded based on the following criteria: (i) overlapping data, (ii) reviews, letters, editorial articles, or meta-analysis, and (iii) incomplete data regarding genotype distribution in cases and controls.Two researchers independently extracted data from each included study and any discrepancies were resolved through discussion. The following information was extracted from each study: first author, year of publication, disease categories, age of controls, genotyping method, sample size of cases and controls, genotype distribution in cases and controls.The methodological quality of included studies was independently evaluated by two researchers according to the Newcastle–Ottawa Scale (NOS) . The NOSP<0.05 considered as a deviation. Dichotomous data are presented as OR with 95% CI. Heterogeneity was assessed using the Q test and quantited by the I2 test. If there was no heterogeneity (P>0.1 or I2 < 50%), a fixed-effects model (Mantel–Haenszel method) was used to estimate the pooled OR; otherwise, a random-effects model (Mantel–Haenszel method) was utilized. For the PvuII -397T/C polymorphism, the allele (C compared with T), dominant (CC+CT compared with TT), recessive (CC compared with TT+CT), homozygous (CC compared with TT), and heterozygous model (CC compared with CT) were evaluated. For the XbaI -351A/G polymorphism, the allele (G compared with A), dominant (GG+GA compared with AA), recessive (GG compared with AG+AA), homozygous (GG compared with AA), and heterozygous model (GG compared with GA) were evaluated. The Z test was conducted to estimate the statistical significance of summary results, with P<0.05 indicating statistical significance. Sensitivity analysis was conducted to assess the stability and quality of the pooled results. The Egger’s test was conducted to evaluate publication bias and in order to assess the potential impact of publication bias on the OR estimate, we estimated an OR adjusted for publication bias using the trim-and-fill method, which imputes potentially missing studies to achieve symmetry in the funnel plot [P<0.05 indicated statistical significance. All statistical analysis was performed using Stata 12.0 software . Statistical power analysis was performed using Power and Precision V4 software  to verify whether the included studies and meta-analysis could offer adequate power (≥80%).The Hardy–Weinberg equilibrium (HWE) was calculated in the control groups using the chi-squared test, with The selection process of eligible articles is shown in ESR1 gene and susceptibility to severe pre-eclampsia.The seven selected studies , which iIn total, five studies ,11,14,15P=0.017) (P=0.008)   and GG cP=0.008) ; other cIn total, we identified three studies ,11,14, cIn one study, it is unclear whether the pre-eclampsia patients were mildly or severely affected , so we pP=0.047) (P=0.921), however, the presence of publication bias did not influence the OR estimate and overall result . The subsequent trim-and-fill method showed that publication bias had little impact on the stability of the results.To identify publication bias, the Egger’s test was performed on each model of the ESR1 gene PvuII and XbaI polymorphisms. The results suggest that there was evidence for publication bias in the GG compared with AA model of the -351A/G XbaI (rs9340799) polymorphism and mild pre-eclampsia, as indicated by Egger’s asymmetry test (P=0.047) . Using tWe adopted the statistical power analysis to reassess the available data when an α of 0.05 and β of 0.2 were assigned. The power of each eligible study of the association between the ESR1 -351A/G Xbal polymorphism and severe and mild pre-eclampsia susceptibility ranged from 5% to 89%, 5% to 45%, respectively. In the meta-analysis of the association between the ESR1 -351A/G XbaI polymorphism and severe pre-eclampsia susceptibility, the power analysis suggests that a power of 57% was required to detect a pooled OR for the GG compared with GA model, and 72% for a pooled OR for the GG compared with AA+GA model. The specific power values are summarized in ESR1 gene. The PvuII, T397C polymorphism occurs due to T>C transition in intron 1; while the XbaI, G351A polymorphism is a G>A transition located 50 bp from the PvuII polymorphic site [The familial nature of pre-eclampsia has been known for a number of years, and the identification of novel susceptibility genes is one of many strategies aimed at fully elucidating the underlying biological pathogenetic mechanisms . The genhic site . For thehic site . As estrhic site , the prehic site . Interprhic site . Furtherhic site . FurtherOur meta-analysis suggests that the GG genotype of the ESR1 XbaI polymorphism could be a genetic risk factor for severe pre-eclampsia susceptibility. As HWE is a surrogate to assess study quality, the chi-squared test was conducted to check whether genotype distributions in controls conformed to the HWE rule. However, the genotype distribution of the controls in two studies ,14 signiThe meta-analysis was a powerful way to effectively increase the sample size to provide a more valid pooled estimate. However, the highest statistical power of the pooled result on the overall meta-analysis of the ESR1 -351A/G XbaI polymorphism and severe and mild pre-eclampsia susceptibility was found to be 72 and 12%, respectively, which are less than 80%, i.e. the powers of our findings were unsatisfactory. Owing to the low statistical power, the results may not be sufficiently robust, and any conclusions should be interpreted cautiously. Statistical power depends upon the population effect size, number of studies, and average sample size. As the pooled sample size (only three or four studies) is not large enough, the statistical power to estimate the effect of this locus may be limited, and thus further large-scale and high-quality studies are necessary to validate our findings.There was evidence of publication bias in the GG compared with AA model of the -351A/G XbaI (rs9340799) polymorphism and mild pre-eclampsia, however, this did not impact the combined OR estimate . The subsequent trim-and-fill method showed that publication bias had little influence on the stability of the results. Publication bias of our meta-analysis was attributed to the limited availability of published results, as the number of publications included in our meta-analysis was relatively small.ESR1 gene were analyzed in the present study. It therefore remains unclear whether additional genetic variants contribute to this gene; investigating haplotypes reveal more information regarding the genetic causes of diseases than genotypes and hence would be more influential than single SNPs. Fourth, including only English and Chinese articles may produce publication bias and more eligible studies could be captured if the search was extended. The selection bias caused by language restriction may reduce the robustness of our meta-analysis [The present meta-analysis has some limitations that should be discussed. First, only four studies assessed the ESR1 -351A/G XbaI polymorphism and severe pre-eclampsia susceptibility, and three studies assessed the ESR1 -351A/G XbaI polymorphism and mild pre-eclampsia susceptibility. Hence, the limited study number and small sample size may not offer sufficient statistical power to investigate the association between the ESR1 polymorphisms and the risk of severe or mild pre-eclampsia. As the statistical power may be limited, more large-scale and high-quality studies will be required to evaluate the risk of pre-eclampsia in different ethnic groups and validate the meta-analysis. Second, all the included studies are case–control and all had the limitation of being observational studies, including selection bias and unmeasured confounders. Third, our meta-analysis did not investigate the genetic haplotypes of severe or mild pre-eclampsia, and only single locus SNP -397T/C PvuII or -351A/G XbaI in the analysis ,24. FinaIn conclusion, our meta-analysis suggests that the GG genotype of the ESR1 Xbal polymorphism may be a genetic risk factor for severe pre-eclampsia susceptibility. However, there is no association between the ESR1 gene Pvull and Xbal promoter polymorphism and the risk of mild pre-eclampsia. Due to the limitations mentioned above, well-designed and larger scale studies are necessary to validate our results, and such investigations may explore the roles of the ESR1 XbaI gene polymorphism in the pathogenesis of severe pre-eclampsia.


Thiazide use was associated with reduced fracture risk in case-control studies, but not in cohort studies. The associations demonstrated in case-control studies might be driven by inherent biases, such as selection bias and recall bias. Thus, thiazide use may not be a protective factor for fractures.The association between thiazide use and fracture risk is still controversial. We conducted an updated meta-analysis on the association between thiazide use and fracture risk. We systematically searched PubMed, Embase, and Cochrane library databases for all types of human studies, including observational and experimental studies that were published up until July 2019. We also manually searched the reference lists of relevant studies. The pooled relative risks (RRs) with 95% credible interval (CrI) were calculated using a Bayesian hierarchical random effect model. A total of 19 case-control  and 21 cohort studies  were included in this meta-analysis. The pooled RR for fractures associated with thiazide use was 0.87 (95% CrI: 0.70–0.99) in case-control and 0.95 (95% CrI: 0.85–1.08) in cohort studies. The probabilities that thiazide use reduces any fracture risk by more than 0% were 93% in case-control studies and 72% in cohort studies. Significant heterogeneity was found for both case-control (p < 0.001, I Thiazide diuretics are one of the most common types of antihypertension medications4. There is evidence suggesting that thiazide diuretics reduce urinary calcium excretion5, and stimulate osteoblast differentiation and bone mineral formation6. Although a previous meta-analysis suggested that thiazide use was associated with reduced fracture risk9, results of individual studies are still inconsistent, ranging from positive to negative effects19. In addition, two previous meta-analyses were published over a decade ago8, and the most recent meta-analysis that was published in 2018 was limited to only prospective cohort studies9. Therefore, an updated meta-analysis that is inclusive of all types of study designs is warranted. We conducted a Bayesian meta-analysis on the association between thiazide use and fracture risk as it uses a probabilistic approach to make clinically relevant decisions in the face of uncertainty. For example, using the Bayesian method, we can determine the probability that thiazide use reduces fracture risk by more than 0%, 10% or 20%; this probability is unable to be provided by classical analysis20. Therefore, we utilized an advanced methodology in meta-analysis research to address the much controversial relationship between thiazide use and fracture risk that encapsulates all peer-reviewed publications in the field thus far.Hypertension and osteoporotic fracture are two major public health problems because they result in a substantial financial burden among the elderly as well as considerable increases in morbidity and mortality21. We systematically searched PubMed, Embase, and Cochrane library databases for all types of human studies, including observational and experimental studies that were published up until July 2019. The keywords and medical subject headings (MeSH) used for the search were: “thiazide” OR “Sodium Chloride Symporter Inhibitors” AND “Bone fracture” OR “Fracture” OR “Osteoporosis”. We also manually searched the reference lists of relevant studies. Studies were included in the meta-analysis if they met the following criteria: (a) were original human studies; (b) used thiazide as an exposure; (c) had risk estimates for fracture outcome. When more than one study used the same data, we included the most recent and best quality study in our meta-analysis.This study was undertaken according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA)22. Briefly, the NOS score was assessed using the following items: selection, comparability, exposure, and outcome; a NOS score of 7 or higher is considered as high quality23.Two investigators  independently identified and extracted all potential articles for inclusion. Any disagreement between the above two investigators was resolved by discussing it with the third author (YL). The following information was retrieved from each study: first author’s name, year of publication, the percentage of female participants, sample size, fracture outcome, mean age, country, and fracture risk estimates. The Newcastle-Ottawa Scale (NOS) was used to assess the quality of each individual study26. In classical meta-analysis, we used the DerSimonian-Laird method27 to calculate the pooled risk ratio. In the Bayesian model, the risk ratios (RRs) for all the studies were converted into a logarithmic scale  and R .We synthesized the data using both classical and Bayesian hierarchical random-effects modelsWe identified a total of 959 articles from different electronic databases and other sources. Of these, 633 duplicate articles and 181 irrelevance articles were excluded after reading the title or abstract. Finally, 19 case-control studies and 21 cohort studies were met for inclusion in this meta-analysis : 0.87, 95% confidence interval (CI): 0.76–0.98). We observed moderate heterogeneity between studies . The heterogeneity between studies was significant (p < 0.001, IThe funnel plot of risk ratio versus standard error for the association between thiazide use and fracture risk was shown in Fig. This meta-analysis provides evidence to support that thiazide exposure is associated with a 13% reduction of fracture risk in case-control studies. However, while an inverse association was noted in cohort studies, it failed to reach statistical significance.et al.8 and Xiao et al.9; both studies suggested that thiazide was associated with the reduction of any fracture risk by 14%. However, to the best of our knowledge, our meta-analysis is the first to distinguish a difference in the relationship between thiazide use and fracture risk by study design. We found that there is a null relationship between thiazide use and fracture risk in cohort studies. A recently published meta-analysis also suggested that the effect of thiazide use on fracture risk was weaker in cohort studies9. Although the results from the Bayesian meta-analysis were consistent with that generated from the classical meta-analysis approach, the Bayesian meta-analysis provides additional regarding the probabilities that thiazide use reduces fracture risk by certain percentages. Such information is useful for making clinically relevant decisions about the use of thiazides, and cannot be obtained using the traditional meta-analysis methodology.Our findings were partly comparable with the effect shown in the previous two meta-analyses reported by Wiens 32. In addition, thiazides are associated with an increased level of metabolic alkalosis, which is an inhibitor of bone resorption34. On the other hand, thiazides diuretics could induce hyponatremia, which has a negative impact on the metabolism and integrity of the bone36. In addition, thiazide induced-hyponatremia could have harmful neurological side effects, such as gait disturbances and imbalance, which leads to an increased risk of falls and fractures37.The controversial relationship between thiazide diuretics and fractures involves conflicting mechanisms. On the one hand, thiazide could exert beneficial effects on the bone via decreasing urinary calcium excretion by 25–40%This meta-analysis has several limitations. First, due to the absence of relevant experimental studies in humans, our meta-analysis included only observational studies. A meta-analysis based on observational studies cannot make causal inferences about thiazide use and fracture risk. Second, we observed considerable heterogeneity between individual studies, which might bias our results. Lastly, due to insufficient data from individual studies, we did not evaluate the effect of dose and duration of thiazide use on bone fractures.In conclusion, this meta-analysis included 19 case-control and 21 cohort studies to examine the relationship between thiazide use and fracture risk. Our results suggest that thiazide use was associated with reduced fracture risk in case-control studies, but not in cohort studies. The associations demonstrated in case-control studies might be driven by inherent biases such as selection bias and recall bias. Thus, thiazide use may not be a protective factor for fractures. Randomized clinical trials are still warranted to confirm our findings.


We summarise the work of the Childhood Cancer Research Group,particularly in relation to the UK National Registry of Childhood Tumours(NRCT).The Group was responsible for setting up and maintaining the NRCT.This registry was based on notifications from regional cancer registries,specialist children’s tumour registries, paediatric oncologists and clinicaltrials organisers. For a large sample of cases, data on controls matched by dateand place of birth were also collected.Significant achievements of the Group include: studies of aetiologyand of genetic epidemiology; proposals for, and participation in, internationalcomparative studies of these diseases and on a classification system specificallyfor childhood cancer; the initial development of, and major contributions to,follow-up studies of the health of long-term survivors; the enhancement of cancerregistration records by the addition of clinical data and of birth records. TheGroup made substantial contributions to the UK government’s Committee on MedicalAspects of Radiation in the Environment.An important part of the ethos of the Group was to work incollaboration with many other organisations and individuals, both nationally andinternationally: many of the Group’s achievements described here were the resultof such collaborations. At the timeof Alice Stewart’s retirement in 1974 the Standing Subcommittee on Cancer of theDHSS Standing Medical Advisory Committee recommended the setting up of a nationalregistry of childhood malignancies. The Childhood Cancer Research Group (CCRG) wasestablished in 1975 within the department of Sir Richard Doll, the Regius Professorof Medicine. The first director was GJ Draper, and he was succeeded in 2002 by MFGMurphy. The remit of the CCRG included maintaining what became the National Registryof Childhood Tumours (NRCT) and conducting research into the epidemiology of thesediseases; from the outset it was assumed that the group would be involved incollaborative studies.This paper is one of three summarising research on childhood cancerdone mainly in Oxford over six decades starting in 1954. The main intention is tosummarise the history and achievements of this research. It is our hope that thesepapers will also serve to draw attention to the availability of the very substantialresearch resources accumulated over this period. The firstpaper2The present paper is one of two describing the genesis andachievements of the work done by the CCRG. It is confined to research studies basedon, or including, data from the NRCT or the OSCC. A large part of the work of theGroup was concerned with studies investigating carcinogenic effects of ionisingradiation; this is dealt with in the accompanying paper by Kendall etal.http://www.nature.com/bjc). This includes a series of papers resulting from participation inthe Inter-regional Epidemiological Study of Childhood Cancer, a case–control studyinvolving cases from three regions of England.Other CCRG work on childhood cancer is included in the list ofreferences on the journal website , birth registration records,information on congenital anomalies and genetic conditions, detailed pathologyrecords and information from clinical trials. In addition, for a large sub-group ofthe cases, a control series was established, with birth registration records,initially for one control per case, matched by sex and by date and area of birthand, for England and Wales, provided by the Office for National Statistics (ONS).Similar controls were also obtained for Scotland. The resulting database became aresource for geographical and other case–control studies (e.g. on exposure toelectric and magnetic fields (EMF), gamma radiation and radon). To improve the powerof later studies two matched controls per case were selected for cases born from2000 onwards. By 2014, when the CCRG closed, information had been collected forabout 57,000 cases born and registered in the UK, 1962–2010, and for about 73,000controls.3which confirmed that the incidence in the nearby village of Seascale was well beyondnormal expectation. This enquiry led to the establishment of a Government advisorycommittee—the Committee on Medical Aspects of Radiation in the Environment (COMARE).A discussion of the work of the CCRG in relation to this committee is included inthe accompanying paper by Kendall et al.2Much of the work of the Group was concerned with epidemiologicalstudies and with studies of the natural history and pathology of particular tumours.A major component of the early epidemiological work consisted in studying thepossible carcinogenic effects of ionising radiation. Initially, this was a responseto a television programme broadcast in 1983—“Windscale, the Nuclear Laundry” thatidentified a larger than expected number of cases of childhood leukaemia in thevicinity of this nuclear installation, subsequently renamed and better known asSellafield. The excess was sufficient to cause alarm and to trigger a majorenquiryThe CCRG worked with many other groups and individuals, bothnationally and internationally. Two very long-term collaborations were especiallyimportant. The first was with other cancer registration organisations in the UK,both nationally—the ONS, previously the Office for Population Censuses and Surveys(OPCS)—and regionally, particularly the regional Childhood Tumour Registries; theNRCT was a member of the UK Association of Cancer Registries and latterly of theNational Cancer Intelligence Network. The second major collaboration was with theCCLG  and with individualpaediatric oncologists and other clinicians.2 Record linkage was also used to identify childhoodcancer survivors who later underwent cardiac transplantation as describedbelow.The inclusion of birth records and in particular parental informationin the NRCT made it possible, using record linkage techniques, and in collaborationwith organisations holding other datasets, for the CCRG to investigate relationshipsbetween childhood cancer and putative aetiological factors. Studies of parental ageand sibship position and of assisted reproductive technology are described below;studies of parents potentially exposed to ionising radiation are described inKendall et al.4and the monograph edited by Stiller.5 Detailed incidence data were also published ina paper describing the methodology of the NRCT.6The CCRG carried out a series of analyses of childhood cancerincidence and survival based on the NRCT, and using standard classificationsystems appropriate to childhood tumours. The most comprehensive national data onincidence, survival and mortality were published in the OPCS volume by Draper etal.7and Stiller.5 In the latter it was concluded that in Britainthere was a total increase in recorded incidence of 38% over the 35-year period1966 to 2000. The increase was seen, in varying degrees over all the maindiagnostic groups of childhood cancers; the authors commented that “Theseincreases in recorded incidence do not necessarily represent real changes inrisk …” but suggested that the change was real at least for acute lymphoblasticleukaemia (ALL) and malignant melanoma. In a study that documented increasingtime trends in recorded incidence of subtypes of childhood leukaemia, Kroll etal.8noted that two separate influenza epidemics each coincided with a small peak inchildhood ALL. Kroll et al.9 validated the completeness of NRCTascertainment for cases of childhood cancer diagnosed during 2003–04. Kroll etal.10 suggested that improved completeness ofdiagnosis and registration “were plausible explanations for most of the changesin recorded incidence” but that “the possibility of some real increases shouldnot be ruled out”.Time trends in incidence were analysed in Draper etal.4 This was followed by detailed studies ofsurvival trends among children diagnosed during 1971–8511 and1980–91,12 the latter including projected long-termsurvival estimates for recently diagnosed patients using a rudimentary form ofthe period approach avant la lettrecf.13 The 2007 monograph included survival analysesfor all children diagnosed during 1966–2000.5 These studies documented theremarkable increase in survival rates from childhood cancer between the 1960s,when five-year survival was below 30%, and 30 years later, when it exceeded 75%.Stiller et al.14 demonstrated how changes in population-basedsurvival for a wide range of childhood cancers paralleled those reported fromthe relevant clinical trials during 1978 onwards.The 1982 report included the first results on nationalpopulation-based survival for all the principal diagnostic groups, coveringchildren diagnosed to the end of 1974.15 applied new methods to define the proportionof children with leukaemia who appeared to be cured—i.e. who, as a group,eventually experience no excess mortality compared with the general population.This rose from 25% for those diagnosed in 1971–1975 to 68% in 1991–1995, thoughthe average time since diagnosis at which cure could reasonably be declared alsoincreased, from 11 years for those diagnosed during 1971–1975 to 16 years forthose diagnosed during 1986–1990, perhaps because of late relapse, secondarymalignancy and toxicity from treatment.16 Stiller etal.14 demonstrated how changes in population-basedsurvival for a wide range of childhood cancers paralleled those reported fromthe relevant clinical trialsShah et al.International collaborative studies of incidence and survival aredescribed below.17 in a follow-up study of 10,106 three-yearsurvivors, confirmed the previously reported high risk of second primary cancersamong patients with heritable retinoblastoma and found that for all otherdiagnostic groups combined the number of subsequent cancers was five times thepopulation rate. Hawkins et al.19 studied the offspring of survivors of childhoodmalignant disease and, albeit on the basis of fairly small numbers and shortfollow-up, found no evidence of mutagenic effects of therapy on the offspring andonly previously recognised patterns of inheritance. Much of the subsequent work onthe health of survivors has been carried out by MM Hawkins in the BritishChildhood Cancer Survivor Study (BCCSS), at the University of Birmingham. TheBCCSS cohort was derived from the NRCT and some of the work was done incollaboration with the CCRG.From the foundation of the CCRG the long-term health of survivorswas recognised as a major object of concern, and a research programme was started.Hawkins et al.The very high risk of other forms of cancer in carriers of theretinoblastoma gene is of particular concern; CCRG studies are included in thesection on retinoblastoma below.21 quantified the need for cardiac transplantationamong those affected, and assessed their survival post-transplantation. Takentogether, these studies demonstrate an increasing need for transplantation amongchildhood cancer survivors. This is most marked for survivors of acute myeloidleukaemia, whose increased survival rates have been achieved as a result of highdoses of cardiotoxic anthracyclines.Another particular area of concern is the risk among childhoodcancer survivors of adverse cardiac effects following treatment. Twostudies1 and continued to ascertain cases for a further40 years, included one of the largest sets of data in the world on childhoodcancer in sibs. Papers on familial risks and patterns of occurrence werepublished.23 In the first, as reportedin,1estimates were made of the cancer risks to sibs of cases. It was concluded that,when cases with certain obviously genetic conditions were excluded, the riskthat a sib of a child with cancer would also be affected by cancer below age 15years was double the normal risk. In the absence of known environmental factorsthat could account for this risk, it seems reasonable, for genetic counsellingpurposes, to act on the assumption that this increase is a consequence of ashared genetic background.The NRCT, having inherited data from theOSCC,24 using data from the Nordic countries,concluded that given information on family histories and the ability torecognise syndromes such as Li-Fraumeni, there was no unexplained increase inthe risk to sibs of cases. This does not, however contradict the finding ofDraper et al.;23 the point here is that, although the findingof an increased risk can be explained by information obtained ex post facto, the genetic counselling situation isthat the question of whether a sib of a case has an increased risk will oftenhave to be assessed in the absence of such information. Draper etal.23 calculated the risk to a future sib on thebasis of the information that the proband had cancer but in the absence of priorknowledge about genetic disease in the family; this is the relevant risk forgenetic counselling purposes. Winther et al.24 calculated the risk forsibs excluding families where there is evidence (including that from the secondsib) that there is a known familial syndrome; this is relevant to the questionof whether, after allowing for these known syndromes, there are stillunrecognised genetic factors associated with childhood cancer. Winther et al.25 The rarity of childhood cancers and ofmonozygotic twins means that it has been impossible to give estimates ofconcordance rates except for leukaemia. For this disease Buckley etal.25 estimated that the concordance rate formonozygotic twins is 5%. Chaganti et al.26 and Ford etal.27 have shown that this concordance can, in atleast some cases, be explained by intraplacental transfer of transformedcells.If one member of a twin pair is affected by childhood cancerthere seems to be an increased risk that the co-twin will be affected by thesame disease.28 suggested that this was because a member ofa pair affected in utero may have an increased risk of dying before the twinpregnancy is recognised as such. See also Bithell etal.1and Murphy et al.30 suggested other possible explanations, e.g.the lower-than-average birthweight of twins.Twins are less likely than singletons to develop childhoodmalignant disease. Hewitt et al.33 For some decades, it has been recognisedthat survivors are at risk of bone and soft-tissue cancers. But as follow-upperiods increased it became clear that this risk extends to a wide variety ofother types of cancer. A study of cancer rates among relatives ofcases34 was among the first to document the highcontinuing lifetime risk of other cancers among carriers of retinoblastoma genemutations.Retinoblastoma is of particular interest and concern, mainlybecause of the large proportion (40–45%) of cases that have a well-understoodgenetic origin, and the fact that RB1 gene mutations are associated with othercancers in addition to retinoblastoma. CCRG studies included analyses ofsurvival and the incidence among survivors of second primary tumours at othersites.35 gave estimates of the risk of retinoblastomafor sibs and offspring of cases according to whether or not these probands hadthe heritable form of the disease and whether it was unilateral orbilateral.Draper et al.36 and of childhood cancer risk associated with awide range of anomalies37 and also investigations related to specificsyndromes. The association between hepatoblastoma and polyposis coli was firstreported by Kingston e al.38 who studied 113 cases and estimated on thebasis of this, admittedly small, sample that mothers of children withhepatoblastoma had about 200 times the population risk of polyposis coli—or, ifit is assumed that the association applies equally to fathers, that the risk ofa parent being affected is 100 times the population risk. Threestudies41 reported on survival of children with Downsyndrome and leukaemia, and the NRCT made a major contribution to internationalstudies highlighting the apparent protective effect of trisomy 21 againstneuroblastoma and medulloblastoma.43 Population-based estimates of the risk ofleukaemia and lymphoma associated with neurofibromatosis were calculated byStiller et al.,44 several years before it began to be recognisedthat a large proportion of patients apparently affected with this rarecombination of conditions had what has come to be known as constitutionalmismatch-repair deficiency syndrome. The CCRG was a long-term member of theFactors Associated with Childhood Tumours (FACT) collaboration, based at theInstitute of Cancer Research, whose series of studies has identified severalgenes involved in the development of embryonal and othertumours.52Recording of congenital anomalies and genetic conditions in theNRCT made possible studies of the heritable fraction of childhoodcancer2 Much of this work was done incollaboration with the National Radiological Protection Board, which laterbecame part of the Health Protection Agency and then Public HealthEngland.The CCRG carried out a series of studies on the possible effectsof ionising radiation on the incidence of childhood cancer; these are describedin detail in the associated paper by Kendall et al.53 reported an increasedleukaemia risk for children with a residential address at birth within 600 m ofa high-voltage powerline. Subsequent analyses demonstrated that this leukaemiarisk declined over time from a significant increase in the 1960s and 1970s to noincrease or a non-significant decrease in the more recentdecades.54 These unexpected findings relating to distancefrom powerlines are not consistent with the existing body of evidence onmagnetic fields56 and have prompted replications in othercountries59 that did not show similar distance effects.Further CCRG studies found no significant association between cancer risk andproximity to high-voltage underground cables55 nor could the observedeffects of overhead lines be explained by the corona-ionhypothesis.60 It was, however, concluded that some linkbetween the presence of high-voltage powerlines and socioeconomic or demographicfactors in the vicinity was the most likelyexplanation.61Possible effects of electric and magnetic fields nearhigh-voltage powerlines have been studied in collaboration with John Swanson ofthe National Grid Company. An early study62 prompted directly by the previous UKfindings,54 examined residential distance fromhigh-voltage powerlines in relation to childhood leukaemia risk. This new pooledanalysis found no material association between childhood leukaemia and distanceto the nearest overhead powerline of any voltage, but a small thoughnon-significantly increased risk for children living <50 m from 200+ KVpowerlines. The distinctive features of the previous UK findings were notconfirmed in other countries and no clearer explanation for the increased riskswith distance found in various studies has been revealed.NRCT data on powerlines have been contributed to threeinternational pooled studies of childhood cancer. Two examined magnetic fields,and confirmed the previously reported association for leukaemia but providedlittle evidence of an association for brain tumours). Thethird,64 reported a possible doubling of the risk ofchildhood cancer from intramuscular vitamin K, given to babies to preventvitamin K deficiency bleeding. Two CCRG studies66 were carried out to investigate thissuggestion; these authors concluded that there might be a small effect on theincidence of childhood leukaemia but none for other cancers. Roman etal.67 in a pooled analysis of six case–controlstudies, including the CCRG study, concluded that while “…small effects cannotbe entirely ruled out, our analysis provides no convincing evidence thatintramuscular vitamin K is associated with childhood leukaemia”. And there waseven less evidence of a risk for the combined group of other childhoodcancers.Golding et al.68 in the possibility of therebeing a differential childhood cancer risk in children born following assistedreproduction and in particular in vitro fertilisation (IVF). The NRCT with itscomplete coverage of childhood tumours in Great Britain offered a uniqueresource for addressing this question. Data on children born after IVF treatmentbetween 1978 and 1991 were initially held by the Medical Research Council, andrecord linkage techniques were used to establish the numbers of such childrenwho went on to develop childhood cancer.69 The Human Fertilisation andEmbryology Authority (HFEA) took over responsibility for record keeping forassisted births from 1992 onwards. Although the HFEA was committed to monitoringthe long-term outcomes for children born following assisted reproductioninterventions, legislation had precluded any release of data on individuals. Amajor revision of the Human Fertilisation and Embryology Act in 2008 allowedapproved researchers highly regulated access to individual HFEA records. Incollaboration with colleagues at University College London, a definitive studywas undertaken70 to establish whether children born between1992 and 2008 following IVF type procedures were at an increased childhoodcancer risk when compared to their naturally conceived peers. Overall, childrenborn following these procedures were not found to be at increased risk ofdeveloping childhood cancer. Further regulation required the analysis ofoutcomes for children born following interventions involving donor eggs, spermor embryos to be undertaken separately. However, again, no increased risk ofoverall childhood cancer was found.71 In two papers, analysing cancer incidenceamong 26,692 children who were born after IVF during the years 1982–2005 inSweden an increased risk of Langerhans cell histiocytosis wasfound.73 A collaborative UK study describing theincidence of this condition in children born following assisted reproductionprocedures is nearing completion.There has long been wide interest74 For this study, it was necessary to havetemporary access to the confidential birth records held at OPCS, to enablelinkage be carried out between these records and those held by CCRG. Onlyanonymised records could be released after the linkage process. The moststriking findings from this study of 10,162 matched pairs were that the risk ofALL decreased with increasing parity and increased with increasing parental age;the latter effect was not explained by the association with Down syndrome. Apredicted association between high paternal age and new germ cell mutations inretinoblastoma was also found, though this was not statistically significant.The corresponding result for maternal age was in fact rather more marked, thoughbecause of the strong correlation between paternal and maternal age it was notpossible to disentangle these effects.A case–control study of the relations between childhood cancers,parental age and sibship position was carried out in collaboration withOPCS.77 which, taken as a whole, showed birthweightto be associated with risk for approximately half of all childhood cancers.Increasing birthweight raises the risk most notably of leukaemia, tumours of thecentral nervous system, renal tumours and soft-tissue sarcomas. Associationswere also observed between high birthweight and the risk of neuroblastoma,lymphoma, germ cell tumours and malignant melanomas. By contrast, increasingbirthweight reduces the risk of hepatic tumours. No association was observedbetween birthweight and the risk of retinoblastoma or bone tumours. Results forUS datasets were very similar to those for the UK(NRCT).Clinically recorded birthweight was appended to the birthregistration details of nearly all children born in England and Wales from about1980. This made possible some of the largest studies of intrauterine growth andchildhood cancer risk82 Overall, these studies showed little, ifany, support for the hypothesis that paternal occupational exposure is animportant aetiological factor for childhood cancer. Analysis of leukaemia cases,however, showed some evidence of a positive association with paternaloccupations involving social contact; in addition, higher paternal occupationalsocial class was associated with increased lymphoid leukaemiarisk.78A job exposure matrix approach, based on the details of father’soccupation provided at birth registration, was used to study paternaloccupational exposures for the majority of childhood cancers registered by theNRCT.83 A study in which the NRCTwas linked to clinical trials data was consistent with this gradient beingpartly due to systematic under-diagnosis in less affluent communities in Britainduring the 1980s and 1990s.84 The authors concluded “Under-diagnosis inpoorer communities may have contributed to socioeconomic variation in recordedchildhood acute lymphoblastic leukaemia incidence within Great Britain, andelsewhere” They also stated that their findings for ALL were consistent with the“pre-emptive infection hypothesis”, which proposes that some children withleukaemia die from infection without leukaemia being suspected. The authorsreferred to Stewart85 who seems to have been the first to proposethis hypothesis.There have been various reports of an association betweensocioeconomic status and the incidence of childhood malignant disease. Forchildhood ALL, persistence of a small socioeconomic gradient reported inprevious studies was confirmed by Kroll et al.86 and Baglan Bay, Wales following concern aboutthe incidence of leukaemia and lymphoma in children and youngpeople.87 None of these investigations led to a clearcausal connection with any putative risk mechanism, but they nevertheless haveconsiderable socio-political value for allaying anxieties about environmentalhazards.The CCRG focus on childhood cancer and leukaemia around nuclearinstallations stimulated geographical analyses of the NRCT data, which wereenhanced by accurate geo-referencing of addresses at birth and registration.Reports of aggregations of cases in a number of areas were examined, includingCamelford, Cornwall following a water pollutionincident,The problem of assessing such “clusters” of cases is exacerbatedby the widely believed hypothesis that cases tend to cluster naturally.Considerable effort has been expended on exploring this hypothesis, particularlyin view of the possibility that clustering might be indicative of a contagiousmechanism.88Because of the widespread interest in geographical variation, theCCRG made a standard dataset on childhood leukaemia and non-Hodgkin lymphoma(NHL) available to a number of other research groups, in order for them toexplore a variety of analytical methods for the examination of incidence ratesand the detection of clustering. The results were collated and published in anOPCS volume.89 and further described in a series of papers incollaboration with colleagues in the University ofNewcastle.91 NRCT data were also used for EUROCLUS, acollaborative European study of the issues.92 Certainly there is evidenceof some variation in incidence of childhood leukaemia, but this is largelyrelated to differences between districts with varying socioeconomiccharacteristics (see above). Kinlen’s “population-mixing” hypothesis, linkingrisk of childhood leukaemia to herd immunity, was investigated in a series ofcollaborative studies using NRCT data.96 A recent review of leukaemia clustering isgiven in Bithell.97Comprehensive geographical analyses of NRCT data are reported inCOMARE’s 11th Report98 and the possible hazard of proximity torailway lines99; in the former paper an increase in leukaemiawas detected in urban areas with high levels of inward migration; in the lattera very weak effect of proximity could be attributed to two urban wards, possiblyaffected by the population mixing phenomenon. The possibility of a commonaetiology for different tumours was explored in a further collaborative paper onthe cross-space-time-clustering of places and dates of birth and diagnosis; twolinks involving Hodgkin lymphoma were regarded as marginally significant afterallowing for multiple testing.100Geographical data were later made available to colleagues in theUniversity of Newcastle for collaborative studies of populationmixing101 About half the cases were still at thebirth location at diagnosis and those who had moved had usually not moved far.This is important in interpreting aetiological studies in which exposures to theagents of interest vary with locationThe NRCT has the residential addresses at both birth anddiagnosis for cases. An investigation quantified the relationship betweenresidence at birth and diagnosis102 used NRCT data in an investigation of theprecision of various sources of such location information, particularly in thecontext of studies of non-ionising radiation.Swanson et al.104 and continued with studies of Wilmstumour,106 leukaemia110 and bone tumours111 among others. Many of thesestudies revealed survival advantages associated with treatment at large specialistcentres or inclusion in clinical trials.  Pritchard et al.106 showed that there was arisk of over-treatment for patients with Wilms’ tumour treated outside paediatriconcology centres.Throughout its existence, the CCRG carried out population-basedstudies of aspects of clinical service delivery such as type and caseload oftreating hospital and entry to clinical trials for a wide range of childhoodcancers, usually in relation to outcome. This work began withretinoblastomaWe believe that these studies on the effects of entry to clinicaltrials and on centralisation of clinical care have had an impact on clinicalpractice.114 and ovary,115 cardiactumours,116 leukaemia,117carcinomas,118 soft-tissue sarcoma120 and NHL.122As already indicated, one of the most important aspects of theCCRG’s work was the extensive series of collaborative studies with otherorganisations and individuals. Many analyses based wholly or in part on NRCT datawere carried out by, or in collaboration with, the UKCCSG/CCLG, individualclinicians and pathologists, and clinical trial organisers. Sets of records or theresults of analyses relating to particular diagnostic groups or to groups of casesin particular hospitals or areas were made available to paediatric oncologists andother research workers. The subjects of these studies included tumours of thekidney123A national registry of paediatric myelodysplasia was set up andanalysed in collaboration with colleagues at Great Ormond StreetHospital.Most of the collaborative studies referred to above involvedepidemiologists and clinicians within the UK. There were also two groups ofinternational collaborative studies:125 with a third volume in preparation, a seriesof papers on the same topic128 and successive editions of the standardclassification system for childhood tumours.131 Other IARC projects in which CCRG played animportant part were a study of leukaemia incidence following the Chernobylaccident133 and the Automated Childhood CancerInformation System (ACCIS), concerned with cancer incidence and survival amongchildren and adolescents throughout Europe, for which the initialpaper134 was followed by a monographic special issueof European Journal ofCancer135 containing 19 papers, nine of theminvolving CCRG authors.The CCRG initiated, and was a leading collaborator in, the study“International Incidence of Childhood Cancer” based at the International Agencyfor Research on Cancer (IARC), which has resulted in twomonographs141 Most recently, the NRCT contributed data andexpertise to the CONCORD study of worldwide cancersurvival.143The CCRG participated in all analyses of childhood cancersurvival within the Europe-wide EUROCARE collaboration.International Incidence of Childhood Cancer and the suite of papersbased on those data paved the way for CCRG involvement in other internationalcollaborations both at IARC and elsewhere.The NRCT, and work of the CCRG based upon it, has been central indeveloping understanding of patterns and trends in childhood cancer and itstreatment. Work on retinoblastoma has been a particular focus. Studies of familialfactors have also been important. A major area of work has been aetiologicalstudies where important contributions have been made to our understanding of risksassociated, for example, with both ionising and non-ionising radiation and withassisted reproductive technology. “Clusters” of childhood cancer have attractedgreat public interest and work based on the NRCT has been vital in investigatingthis topic. The success of the first volume of The main research resource developed by the Group was thecomputerised NRCT database and a large series of associated paper records anddigitised images of many of these. Most of the work described above was based onthese electronic and paper records. In addition, in the course of both the OSCCand the routine data collection of the NRCT a series of records relating to familyhistories generally, and to sibs and twins in particular has been assembled; thereis in particular an especially large set of pedigree data relating toretinoblastoma. The data are archived; we are attempting to ensure that, subjectto the usual restrictions and permissions concerning access to confidential data, they can bemade available for specified research projects.Another major resource was an electronic database containingbibliographic details of around 16,000 references related to childhood cancer;many of these include electronic links to the actual publications.We hope that one consequence of the publication of this paper andthe two accompanying ones will be to draw attention to the existence of theseresources.Supplementary information is available for this paper at https://doi.org/10.1038/s41416-018-0181-z. This is a file of CCRG Publications on Childhood CancerResearch


Conventional meta-analytic procedures assume that effect sizes are independent. When effect sizes are not independent, conclusions based on these conventional procedures can be misleading or even wrong. Traditional approaches, such as averaging the effect sizes and selecting one effect size per study, are usually used to avoid the dependence of the effect sizes. These ad-hoc approaches, however, may lead to missed opportunities to utilize all available data to address the relevant research questions. Both multivariate meta-analysis and three-level meta-analysis have been proposed to handle non-independent effect sizes. This paper gives a brief introduction to these new techniques for applied researchers. The first objective is to highlight the benefits of using these methods to address non-independent effect sizes. The second objective is to illustrate how to apply these techniques with real data in R and Mplus. Researchers may modify the sample R and Mplus code to fit their data.The online version of this article (10.1007/s11065-019-09415-6) contains supplementary material, which is available to authorized users. The heterogeneity variance τ2 is an absolute index of heterogeneity that depends on the type of effect size. That is, we cannot compare the computed heterogeneity variances across different types of effect size. We may calculate a relative heterogeneity index I2 to indicate what percentage of the total heterogeneity is comprised by between-study heterogeneity  effect sizes, for example, the raw mean difference and the Fisher’s z transformed score, work well even for small sample sizes when the underlying populations are normally distributed.There are two critical assumptions in random- and mixed-effects meta-analyses. First, the sample effect size SEs) on the average effect are generally under-estimated  are nested within these higher-level units. There are several such examples in cross-cultural meta-analyses, where the studies are nested within the countries . The meta-analytic model in Eq.  = T2 is the p × p population heterogeneity variance-covariance matrix that has to be estimated, and Cov(ei) = Vi is the p × p known sampling variance-covariance matrix in the ith study that is computed from the summary statistics ) by using a multivariate meta-analysis.When comparing the univariate and multivariate meta-analyses, Ishak, Platt, Joseph, and Hanley  have argTwo factors may affect the usefulness of multivariate meta-analyses. The first factor is the correlation between the population effect sizes. The presence of a positive (or negative) association between the effect sizes reduces the uncertainty of the estimates of other effect sizes. This is similar to the case of the MANOVA . On the other hand, the three-level meta-analysis allows researchers to study the heterogeneity variances τ2 (and R2) at different levels, whereas the robust variance estimation combines these effects into one single value.Before leaving this section, we have to mention another procedure that is used to handle dependent effect sizes. It is called the robust variance estimation  on asthma and lower respiratory disease are 0.23 (0.05) and 0.30 (0.06), respectively. The estimated heterogeneity variances on asthma and lower respiratory disease are 0.04 and 0.05, respectively. The estimated I2 on asthma and lower respiratory disease are 0.73 and 0.92, respectively.There are a total of 59 studies in the data set “Nam03” in the metaSEM package. Eight of these studies include both asthma and lower respiratory disease, while the remaining studies only include one of these two effect sizes. If we conduct two separate meta-analyses, the average effects  and 0.31 (0.05), respectively. The estimated heterogeneity variances on asthma and lower respiratory disease are 0.07 and 0.05, respectively. The estimated We may take advantage of the multivariate meta-analysis by testing several additional research questions. First, the estimated correlation between the random effects is .96, which suggests that studies with a large effect on asthma tend to be associated with studies with a large effect on lower respiratory disease. Figure χ2(df = 2) = 2.78, p = .25. Therefore, there is no evidence to reject the null hypothesis that the effects are the same in asthma and lower respiratory disease.In a multivariate meta-analysis, we may test whether the average effects on asthma and lower respiratory disease are the same and whether their heterogeneity variances are also the same. Comparing the models with and without these two constraints on the means and variances, the SEs are −0.04 (0.02), p = .01 and − 0.02 (0.01), p = .01, respectively. Their R2 are .59 and .39, respectively. The effect of environmental tobacco smoke is weaker in studies with older participants. Similarly, we may also test whether the regression coefficients on asthma and lower respiratory disease are the same. By comparing the models with and without the constraint on the regression coefficients, the χ2(df = 1) = 0.64, p = .42. Therefore, there is no evidence to reject the null hypothesis that the moderating effect of the mean age of the participants is the same in asthma and lower respiratory disease.We may further conduct a mixed-effects multivariate meta-analysis by using the mean age of the participants as a moderator. The estimated regression coefficients on asthma and lower respiratory disease and their In the above analyses, we used a correlation of .5 to calculate the sampling covariances between the effect sizes of asthma and lower respiratory disease. We conducted a sensitivity analysis using a correlation of 0 and .8. The results were very similar. Therefore, our results are robust to the choices of the correlation in calculating the sampling covariances between the effect sizes.The second example was based on the data set from Stadler, Becker, Gödker, Leutner, and Greiff , Table 1SE) is .42 (.03). The estimated heterogeneity variance and the I2 are .04 and .96, respectively. Based on the three-level meta-analysis, the average correlation (and its SE) is .43 (.03). The estimated level-2 and level-3 heterogeneity variances are .02 and .02, respectively while the estimated level-2 and level-3 I2 are .45 and .51, respectively. The three-level meta-analysis provides more information on how the heterogeneity can be decomposed into the level-2 and level-3 components. The results suggest that the study level can account for more heterogeneity (51%) than the effect size level does (45%).If we ignore the dependence and conduct the univariate meta-analysis, the average correlation (and its χ2(df = 1) = 4.52, p = .03. The average correlation between complex problem solving and intelligence is stronger for studies with a reasoning measure, at .48 (SE = .04), than for those with a general intelligence measure, at .35 (SE = .05).In the dataset, the effect sizes are based on two different intelligence measures . It is of interest to test whether the effects on these intelligence measures are the same. We include the intelligence measure as a moderator in the three-level meta-analysis. By comparing the models with and without the moderator, we find that the change in the chi-square statistics was This paper introduced the problems and preferred solutions for handling non-independent effect sizes in a meta-analysis. Multivariate meta-analyses and three-level meta-analyses can handle different types of non-independent effect sizes. Besides providing valid statistical models to handle non-independent effect sizes, multivariate and three-level meta-analyses allow researchers to address new research questions that cannot be answered in a conventional meta-analysis. In a multivariate meta-analysis, we may compare the average effects or heterogeneity variances across different types of effect sizes. We may also study how the population effect sizes are correlated. In a three-level meta-analysis, we may investigate the heterogeneity variances and explained variances at different levels.A multivariate meta-analysis is usually more challenging to implement because we need to know the correlation between the effect sizes. Many primary studies, however, may not include information on how to estimate this correlation. In contrast, it is easier to implement a three-level meta-analysis because the degree of dependence is estimated from the data. As we have illustrated in the above example, a three-level meta-analysis may also be used to handle different types of effect sizes, namely, the outcome measure in our illustration.In this paper, we simplify the non-independence into either multivariate or nested effect sizes. Then a multivariate meta-analysis and a three-level meta-analysis are used to address the non-independence of the effect sizes. In applied research, the type of dependence is usually more complicated than in cases with either multivariate or nested effect sizes  may be biased. Ideally, we may want to model all types of dependence. However, it is sometimes challenging to do this. Further studies may clarify when it is acceptable to drop or combine the effect sizes to simplify the analyses.The effect sizes may still be non-independent even though each study only contributes to one effect size. For example, Shin  found thBefore closing this paper, it is important to discuss a few issues. First, the selection of effect sizes should be guided by the research questions. Researchers should not blindly include all effect sizes simply because the effect sizes are available. Researchers should carefully define the inclusion and exclusion criteria and use these criteria to determine whether or not the effect sizes should be included.Another issue is the number of effect sizes needed to conduct a three-level meta-analysis. Similar to a standard meta-analysis and multilevel model, the fixed-effects estimates are usually quite stable whereas the stability of the estimated level-2 and level-3 variance components depends on the number of effect sizes for the level-2 and level-3 data. For example, López-López et al.  showed tIn conclusion, researchers have to properly incorporate the dependence in a meta-analysis. The recent development of multivariate and three-level meta-analyses provides a good starting point from which to analyze non-independent effect sizes.ESM 1(PDF 306 kb)


I2 = 40.5%). In addition, exposure to PFOA in early life increased the z-score of childhood body mass index . Accordingly, exposure to PFOA in early life is associated with an increased risk for childhood adiposity. Further research is needed to verify these findings and to shed light on the molecular mechanism of PFOA in adiposity.Some articles have examined perfluorooctanoic acid (PFOA) exposure in early life in relation to risk of childhood adiposity. Nevertheless, the results from epidemiological studies exploring the associations remain inconsistent and contradictory. We thus conducted an analysis of data currently available to examine the association between PFOA exposure in early life and risk of childhood adiposity. The PubMed, EMBASE, and Web of Science databases were searched to identify studies that examined the impact of PFOA exposure in early life on childhood adiposity. A random-effects meta-analysis model was used to pool the statistical estimates. We identified ten prospective cohort studies comprising 6076 participants with PFOA exposure. The overall effect size (relative risk or odds ratio) for childhood overweight was 1.25 : 1.04, 1.50;  The prevalence of obesity has increased worldwide over the past four decades, and is now a global health issue, with over 340 million children and adolescents aged 5–19 being overweight or obese in 2016 [Perfluoroalkyl substances (PFASs) are a class of synthetic and highly stable compounds that are widely applied in protective coatings of food, textiles, furniture, and nonstick cookware ,6,7. Then = 444) observed a significant association between maternal serum concentrations of PFOA and overweight risk [The literature exploring the relationship between exposure to PFOA and adiposity has increased dramatically over the past decade, with a particular focus on early-life exposure. However, the evidence is controversial. A prospective study of Faroese children 5 years of age  independently conducted all studies by title or abstract, and then by a full-text screening. Any discrepancies among the two reviewers reached a consensus by discussion. Studies were included in the meta-analysis when they satisfied all of the following criteria: (1) the study was designed as a prospective cohort study; (2) offspring were exposed to PFOA in utero or in the first year of life (early-life exposure), and maternal serum or plasma concentrations served as proxies for early-life exposure; (3) the outcomes of interest were childhood overweight, obesity or body mass index (BMI); (4) the child’s weight and height were measured when the child was more than 2 years old, because BMI is a commonly used indicator of excess weight in children over 2 years old; and (5) provided risk estimates on the associations, such as relative risks (RRs), odds ratios (ORs), or beta regression coefficients (βs) with 95% confidence intervals (95% CIs), or provided sufficient data to allow adequate estimation of the RRs/ORs/βs and 95% CIs. Studies containing overlapping data or that did not report original data, such as reviews, comments, editorials, and meeting abstracts, were all excluded.Two reviewers (P.L. and F.Y.) extracted data from all of the included articles independently, and discrepancy was resolved by a third reviewer (Y.W.) for quality assurance and quality control. We collected details of selected articles, primarily including first author, study design, baseline time, location, outcome measurements, measurement timing, duration of follow-up, sample matrix and adjusted variables. When outcomes were measured at different follow-ups, we extracted the data for the longest follow-up time. The authors were contacted to request crucial information data not provided in the selected articles. If no response was received from the authors within one month, the data were excluded from analysis.The quality evaluation of included studies was performed using the 9-star Newcastle–Ottawa scale, which includes 3 categories: selection , comparability , and outcome  for cohort studies. Each category that meets the criteria can be awarded a star, except for the comparability category (which awards a maximum of 2 stars). A score of seven or more points was considered as high-quality study .β regression coefficients were used for childhood BMI z-score, and RRs and/or ORs were used for childhood overweight. When different β/RR/OR values for different percentiles, tertiles, or quartiles of PFOA instead of continuous trends were provided in included studies, the strongest association with the outcome (worst-case scenario) was selected [The risk estimates on the association between early-life exposure to PFOA and childhood adiposity were pooled using meta-analysis. The selected . The effselected .I2 statistic and was represented on a scale ranging from 0% to 100%. I2 values of 25%, 50%, and 75% were considered as low, medium, and high heterogeneity, respectively [p < 0.05 hints publication bias. Egger’s test detects funnel plot asymmetry by determining whether the intercept deviates significantly from zero in the regression equation based on the standard normal deviate and precision of each study included in the meta-analysis [α of 0.05 was considered statistically significant.Heterogeneity was measured with the ectively . A randoectively . Subgrouanalysis . All staz-score across childhood but did not report the outcomes of interest such as childhood overweight, obesity or BMI. In addition, hand searching the reference lists of the 9 included articles identified 1 additional study, resulting in a total of 10 articles involving 6076 participants that were retained for quantitative synthesis in the meta-analysis [Our search retrieved a total of 1576 articles, of which 1433 were retained after manual removal of 143 duplicates. After sifting titles and abstracts, we retained 10 records for full-text screening, resulting in 9 full-text peer-reviewed articles retained for data extraction; 1 study was further removed due to no outcomes of interest . Specifianalysis ,30,31,32The characteristics of the 10 cohort studies are detailed in z-score, including the International Obesity Task Force standards [z-score ≥ 85th percentile [z-score ≥ 85th percentile [2 [2 and 17.75 kg/m2, respectively [As for outcomes assessment, different standards were used to calculate BMI tandards , the 200tandards ,28, longtandards , or the tandards ,29,31,32rcentile ,29, 2 dentile [2 , 1 definectively , and 1 dectively .Among those 10 studies, nine analysed the influence of early-life exposure to PFOA on childhood BMI ,30,31,32I2 = 40.5%). When the studies stratified analysis by the effect size (relative risk vs. odds ratio), a significant correlation between early-life exposure to PFOA and childhood overweight risk was observed in group of relative risk (RR) , while the odds ratio (OR) group had a slightly higher assessment but no significant risk for childhood overweight . The pooled effect sizes for males and females were selected for all studies, with the exception of Halldorsson et al. [One of the 10 cohort studies reported overweight and obesity data, respectively. Overweight and obesity data were combined as overall overweight data because it is considered to be more effective . As shown in n et al. , where oβ for childhood BMI z-score after early-life exposure to PFOA. One reported separate β for males and females, and thus was entered as two studies [z-score of childhood BMI with a moderate heterogeneity .Among those 10 studies, nine provided pooled  studies . As showz-score of childhood BMI . A significant association between early-life exposure to PFOA and childhood BMI z-score was observed among the studies performed in Europe , but not in Northern America  and Asia . Subgroup analysis of maternal parity revealed a significant positive association between early-life exposure to PFOA and childhood BMI z-score in the group adjusted by maternal parity . Within the group by gender, PFOA exposure increased the z-score of childhood BMI in females , but not males , and neither achieved significant levels. In addition, the subgroup of birth weight was evaluated, and PFOA exposure could statistically significantly increase the z-score of childhood BMI in the group that was not adjusted by birth weight   .β = 0.07, 95% CI: 0.01, 0.14; I2 = 14.4%). In brief, the results of the sensitivity analysis did not alter the overall association of PFOA exposure and childhood adiposity.For childhood BMI, sensitivity analysis showed that the confidence intervals did not overlap the null in the eleven possible combinations . The numEgger’s test = 0.054) (As shown in = 0.054) . Howeverα or γ, which act as master lipid regulators to modify lipid homeostasis [Currently, excessive food intake, inadequate exercise, endocrine-disrupting chemicals, and possible gene-environment interactions are major determinants of obesity. Evidence has shown that the critical window of exposure to endocrine-disrupting chemicals is either in the prenatal or neonatal period . PFOA iseostasis ,39. Thireostasis . Howeverz-score of childhood BMI. Thus, the results of this meta-analysis demonstrated a positive overall association of PFOA exposure in early life with childhood adiposity.To the best of our knowledge, this meta-analysis is the first study to evaluate the impact of early-life exposure to PFOA on childhood adiposity. A final total of ten prospective cohort studies were included. The results of our meta-analysis showed that early-life exposure to PFOA had statistically significant correlation with increased risk of childhood overweight and slightly increased the CD-1 mice exposed to PFOA for 17 days (0.01 to 5 mg/kg body weight) during pregnancy showed that low-dose exposures (0.01–0.3 mg/kg/d) significantly increased body weight in mid-life of offspring . The stuz-score of childhood BMI in the adjusted maternal parity group , whereas the association was non-significant in the non-adjusted maternal parity group . Therefore, this result suggested that maternal parity had a potential effect on childhood obesity, and the positive association between early-life exposure to PFOA and the z-score of childhood BMI might more truly express that association, further confirming the reliability and robustness of our results.Potential bias should be considered in the meta-analysis. Although the meta-analysis of early-life exposure to PFOA and child BMI showed moderate heterogeneity, we performed subgroup analyses to assess potential bias. Maternal parity is a known predictor of PFAA exposure , and beiz-score of childhood BMI, and both reached a statistically significant level . The results are broadly consistent with a previous study that demonstrated that PFOA concentrations were highly correlated between prenatal and neonatal periods [Quantification of PFOA in included studies was carried out in different matrices, 2 in postpartum maternal serum ,27, 4 in periods .This meta-analysis has several strengths. First, considering the uncontrollability of pollutant exposures and the ethical limitations on conducting controlled trials with pollutants in humans, cohort design is considered the most feasible method. One of the inclusion criteria of our meta-analysis was that the design type of the literature be a cohort study. Second, findings were consistent across multiple outcome types , further supporting the robustness of the findings.z-score of childhood BMI in the female group was slightly elevated but not statistically significant , while the male group showed the opposite trend . This is probably due to the low number of related studies, the included studies were not sufficient to verify gender differences. Future study on this matter should address gender difference with appropriate design and larger sample size. Second, the study was based on the available data, and non-English articles were excluded, thereby making publication and selection biases likely inevitable. Third, it was reported that the distinctions among the OR and RR could be ignored when the outcome is rare [z-score varied across the studies, and the definition of overweight or obesity also varied, which may result in imprecise estimates of association between PFOA exposure and childhood adiposity. Fifth, birth weight is a known predictor of offspring BMI [β = 0.16, 95% CI: −0.05, 0.36, non-adjusted, β = 0.10, 95% CI: 0.03, 0.17). The association between PFOA exposure and childhood adiposity may be exaggerated due to inconsistent birth weight baseline. Sixth, most of the research comes from Europe  and North America , with one study from Asia (Taiwan). There was a significant association between PFOA exposure and childhood adiposity in Europe , but not in Northern America  and Asia , and these results suggested differences between ethnicities. Due to the lack of African data in our study, the findings may not be globally universal. Seventh, no consistent trend was observed after a preliminary examination of the dose-response trend for studies that met the dose-response meta-analysis conditions. We did not conduct dose-response meta-analysis, given the variability between different exposure and reference groups in different studies. In addition, to reduce the impact of confounding factors, we extracted data reported from adjusted models, which means that our results may be susceptible to over-adjustment.There are also several limitations in this meta-analysis. First, one limitation is the study design of the included articles. Sex-specific differences associated with weight gain have been observed for PFOA in some studies. As mentioned earlier, the animal study on CD-1 mice reported higher body weight in pospubertal female offspring after low-dose prenatal exposures to PFOA . Human s is rare . Clearlyring BMI , but onlWith the increasing prevalence of childhood obesity in recent years, effective treatments have not been found, and prevention remains crucial. Exploring the association between PFOA and childhood obesity may help researchers further understand the consequences of early-life exposure to PFOA on offspring obesity and may also provide effective prevention or treatment measures for childhood adiposity, such as reducing risk of exposure to PFOA through early education. Thus, an improved understanding of the association between early-life exposure to PFOA and child adiposity has important implications for public health.In summary, limited to prospective cohort studies, our meta-analysis indicated a statistically significant association between early-life exposure to PFOA and childhood adiposity. However, the results may be affected by other factors, such as we cannot rule out the possibility that these associations were caused by accumulation or synergy of multiple chemical exposures or by potential residual confounding factors. Future research with prudent design is warranted to verify these findings.


Primary open angle glaucoma (POAG) and normal tension glaucoma (NTG) cause irreversible blindness while current medications cannot completely inhibit disease progression. An understanding of immunopathogenesis is thus a keystone to develop novel drug targets and genetic markers are still required for early diagnosis. Toll-like receptor 4 (TLR4) is an essential player in inflammation in various diseases. However, the TLR4 polymorphisms have not been completely elucidated in both types of glaucoma. The aim of the present study was to identify the association between TLR4 polymorphism and glaucoma (POAG and NTG) via the use of a comprehensive review and meta-analysis. The relevant studies were collected from PubMed, Excerpta Medica Database (EMBASE), and Web of Science to identify eight included articles, assessed for quality by a modified Newcastle-Ottawa Scale (NOS) for gene association study. A meta-analysis was applied to calculate the pooled odds-ratio and 95% confidence intervals (CIs) to evaluate the association between TLR4 polymorphism and glaucoma. The results revealed that TLR4 rs1927911 A/G, rs12377632 C/T, and rs2149356 G/T significantly decrease the risk of POAG and NTG in allele contrast models 0.71-, 0.71-, and 0.67-fold, respectively. Moreover, rs4986790 A/G and rs4986791 C/T showed a stringent association with POAG in allele contrast, heterozygous, recessive, and overdominant models. In conclusion, this meta-analysis represented a significant correlation between TLR4 polymorphisms and both types of glaucoma suggesting that TLR4 might be involved in the pathogenesis of glaucoma and may be applied as a genetic marker for disease screening. Glaucoma, characterized by retinal ganglion cell (RGCs) death and optic nerve fiber loss, is a common optic neuropathy which is the second-highest cause of blindness worldwide ,2. It waThe emerging roles of inflammation and immunity were suggested as a paradigm shift in glaucoma pathogenesis. A gain-of-function mutation of pro-inflammatory gene TBK1-tumor necrosis factor (TNF) receptor associated factor NF-κB activator (TANK) binding kinase1 associated with NTG ,8. In a In this decade, advanced sequencing technology has resulted in personal medicine and biomarker development. To achieve this aim, gene association studies are required as personalized information to predict the disease outcomes. Single nucleotide polymorphisms (SNPs) in the TLR4 gene were associated with both infectious diseases and non-communicable diseases . To dateTo identify the related article, the searching, dependent on electronic literature databases, was performed in PubMed, EMBASE, and Web of Science. The search terms were updated until 4 December 2018 and used the following keywords: (variants OR genetic polymorphisms OR genetic polymorphism OR genotyping OR SNP OR SNPs OR SNP OR SNPs OR polymorphisms OR polymorphism OR nucleotide polymorphism OR gene variation OR haplotype) AND  AND (POAG OR POAG OR NTG OR NTG). Two independent investiators searched and subsequently screened the titles and abstracts to identify eligible articles based on inclusion and exclusion criteria. The investigator entered into a discussion to resolve disagreements to achieve the final consensus. An additional publication was considered via related article screening.Selected publications included in our study were: (1) case-control or cohort study on the association of TLR4 polymorphisms and POAG; (2) odds-ratio (OR) or relative risk with a 95% CI; (3) human study, and (4) providing information for all genetic models. The exclusion criteria were: (1) abstract, case-report review, and systematic review; (2) other TLRs study; (3) non-relevant study, and (4) insufficient data.The modified NOS for genetic association study was applied for quality assessment of included articles which were evaluated by two independent investigators . The NOSBriefly, the name of the first author, the publication year, the number of cases and controls, genotype distribution of cases and controls, genotyping methods, and case definition were extracted by two investigators independently extracted relevant information from the eligible studies. An external participant was invited as an expert to resolve some disagreements.P<0.05 showing a deviation from HWE. The strength of association between TLR4 polymorphisms and glaucoma was represented as OR with 95% CIs. All allelic models  were examined for association by using an adjusted P-value for multiple testing via the Bonferroni method. The I2 value was considered to assess the heterogeneity amongst different studies. I2 <50% and P>0.05 was considered a homogeneous population. Consequently, the pool OR was combined using the fixed-effect model otherwise the random-effect models were performed when I2 >50% and P<0.05. Moreover, publication bias was tested by funnel plot and Egger’s regression test. Sensitivity analysis was considered to evaluate the stability of the meta-analysis result and root of heterogeneity by all studies removing one by one. The statistical analysis and meta-analysis were carried out using MetaGenyo ,,27–34], All studies were of case-control design and published between 2008 and 2017. The case definition of POAG was based on clinical manifestation. Briefly, the patient has VF defects or abnormalities, no history of angle closure or secondary causes. Only one study in NTG applied IOP <24 mmHg which has a difference from two studies (IOP < 22 mmHg) .One study performed genotyping using PCR-sequencing and eight studies assessed using real-time PCR. The genotype details are presented in P=0.02, rs12377632; OR = 0.78, P=0.02, rs2149356; OR = 0.71, P=0.02, homozygous model, rs1927911; OR = 0.68, P=0.01, rs12377632; OR = 0.65, P=0.00, rs2149356; OR = 0.62, P=0.02, recessive model, rs1927911; OR = 0.69, P=0.02, rs12377632; OR = 0.70, P=0.04, rs2149356; OR = 0.64, P=0.03). An Egger’s test and funnel plot suggested that no publication bias exists in these genetic models . Altogether, the results suggest that the TLR4 polymorphism  is related to glaucoma in both POAG and NTG.To investigate the potential association of other TLR4 polymorphisms with glaucoma (POAG and NTG), three studies about rs10759930 polymorphism (946 cases and 914 controls), three studies about rs1927914 polymorphism (946 cases and 914 controls), four studies about rs1927911 polymorphism (1133 cases and 1023 controls), four studies about rs12377632 polymorphism (1133 cases and 1023 controls), four studies about rs2149356 polymorphism (1133 cases and 1023 controls), four studies about rs11536889 polymorphism (1133 cases and 1023 controls), four studies about rs7037117 polymorphism (1130 cases and 1144 controls), and three studies about rs7045953 polymorphism (946 cases and 914 controls) were enrolled for meta-analysis resulting in three positions  showing significant association with glaucoma  in five P =0.01, rs4986791; OR = 0.41, P=0.01) , recessive model , and overdominant model   , heterozP =0.03) .2 > 50%, P<0.05). Therefore, we employed stratification by using the glaucoma types (POAG and NTG). All positions were associated with POAG in the recessive model .Since the stratification by using the glaucoma types still showed heterogeneity, we conducted a sensitivity analysis to reveal the influence of each study on the pooled OR of glaucoma. The sensitivity analysis of the association between SNPs  and glaucoma was performed in the allele contrast model and shown in Our meta-analysis was based on systematic collected studies relating to TLR4 gene polymorphisms and glaucoma. To our knowledge, this is the first study that attempted to combine the POAG and NTG studies because these two types of glaucoma are overlapping in some factors ,6. Interin silico analysis showed that rs2149356G is able to form miRNA which targets autophagy genes [in vitro functional assay to prove this hypothesis.Besides the combined analysis, we conducted a subgroup analysis by using types of diseases (POAG or NTG). Surprisingly, rs1927911, rs12377632, and rs2149356, which exist in an intron, were still associated with POAG. However, the functions of polymorphisms have not been examined. Only rs1927911 was associated with NTG. It is possible that there are functions relating to RNA stability and regulation, resulting in alteration of protein expression because several translated regions or a part of intron might relate to miRNA . MoreoveAdditionally, coding polymorphisms of TLR4 rs4986790 A/G (Asp299Gly) and rs4986791 C/T (Thr399Ile), for which both mutations exist in a co-segregation, were also robustly associated with POAG. Taken together, this implied that the TLR4 might play an essential role in POAG pathogenesis. Although the exact functions of these mutations are underinvestigated, there is evidence indicating that these mutations enhance cell death in hepatic stellate cells .Moreover, Asp299Gly and Thr399Ile impair the ability to respond to TLR4 ligands in colorectal cancer cell lines . TherefoTLR4 activation plays a role in innate immunity and subsequently the adaptive immune response . Current+ T cells enter into the retina and cause neurodegeneration in the glaucoma model. The CD4+ T cells are primed by HSP of normal flora and subsequently crossreacted with mouse or human HSP, TLR4 ligands [In the recent year, there is new evidence suggesting that CD4 ligands . Moreove ligands . In addi ligands ,58–60. HThere are limitations which appeared in this meta-analysis and these should not be ignored to improve validity and reliability. First, the sample size should be expanded. Second, a relative descent is biased in the Japanese population. The other population descents are still required, especially for the African, Caucasian, and Han Chinese populations. Moreover, additional factors may affect the results, such as co-segregation events whereby our meta-analysis cannot assess the haplotype between TLR4 polymorphisms and glaucoma. At last, the natural history of IOP is not simplified by a single measurement. However, several included articles, which were enrolled in our meta-analysis, did not mention in this regard. It is well known that the IOP is fluctuation over 24 h . TherefoIn conclusion, the present study demonstrated that there are associations between TLR4 rs1927911, rs12377632, and rs2149356 and glaucoma (POAG and NTG), while rs4986790 A/G and rs4986791 C/T strongly decrease the risk of POAG suggesting that TLR4 may play a role in glaucoma pathogenesis, which should be classified as a neuroinflammatory and autoimmune disease, and should be considered as a genetic marker for POAG. Previously, the role of TLR4 in POAG was exploited and suggested that it may save as a potential therapeutic strategy . HoweverThe data for meta-analysis are available in


Previous studies have demonstrated epidemiological evidence for an association between cancer and the development of new-onset atrial fibrillation (AF). However, these results have been conflicting. This systematic review and meta-analysis was conducted to examine the relationship between cancer and the risk of developing atrial fibrillation.  PubMed and Web of Science were searched for publications examining the association between cancer and atrial fibrillation risk published until June 2017. Adjusted odds ratios (ORs) or hazard ratios (HRs) and 95% CI were extracted and pooled. p < 0.00001; I2 = 67%, p=0.02). The risk of atrial fibrillation was highest within 90 days of cancer diagnosis  and this risk diminished with time.  A total of five studies involving 5,889,234 subjects were included in this meta-analysis. Solid cancer patients are at higher risk developing atrial fibrillation compared to noncancer patients (OR 1.47, 95% CI 1.31 to 1.66,  The risk of AF was highest within 90 days of cancer diagnosis. We should take into account the increased risk of atrial fibrillation development and, after this, study the embolic risk and potential indication of oral anticoagulation. Atrial fibrillation (AF) is the commonest cardiac arrhythmia observed in clinical practice and is associated with significant morbidity and mortality globally. Its prevalence is increasing in part due to an aging population , from ∼0This meta-analysis of observational studies was in accordance with the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) statement .Two researchers (M. Y. and X. F.) systematically and independently searched the relevant studies from the following databases: PubMed (until June 2017) and Web of Science (from 1986 through June 2017). We used the following key words: “cancer,” “carcinoma,” “tumor,” and “atrial fibrillation.” The reference lists of the articles, conferences, and editorials were used to identify further relevant studies.All observational studies reported on the epidemiological evidence for an association between solid cancer and AF were included in this meta-analysis. Because the aim of this meta-analysis is to investigate whether cancer patients are at an increased risk of developing AF in general population, study population that involved the patients who underwent surgery and chemotherapy were excluded. The inclusion criteria were articles (1) published in English language, (2) on human subjects, (3) that were case control, prospective or retrospective cohort study, (4) reporting the odds ratios (ORs) or hazard ratios (HRs) and the corresponding 95% confidence intervals (CIs) or data required for their calculations were provided, (5) assessing the association between AF and cancer, not survival of cancer or complication of AF. Where two articles with overlapping data were included, the articles with higher subjects were included.Two researchers (M. Y. and X. F.) independently screened the titles and abstracts of the studies. Potential eligible studies were retrieved using the relevant inclusion criteria mentioned previously. Any disagreements or indeterminations between the two researchers were resolved through discussion or consultation with a third researcher (T. L.).Hazard ratio (HRs), odds ratio (ORs), and their 95% confidence intervals (CIs) for the association between cancer and AF were extracted from individual articles. If both unadjusted and adjusted ORs/HRs were reported, the adjusted ORs/HRs were preferentially used. And we priority to extracted multivariate adjusted ORs/HRs, not age- or gender-adjusted, to evaluate the risk of AF occurrence. The extracted data of each study included first author's last name, year of publication, geographic location of study, study design, total number of subjects, participants' age and sex, types of cancer, criterion for AF confirmation, follow-up duration, maximum adjusted covariates, HR or OR with 95% CI, and patients with heart failure, hypertension, and diabetic mellitus.The Newcastle-Ottawa Scale (NOS) items, with total score of nine stars, were used to evaluate the quality of cohort or case control studies . We defiI2 statistic, and I2 value of >50% represented moderate to high heterogeneity. The random effects model was used as it is better to explain heterogeneity between studies over the fixed-effects model. Subgroup analyses and sensitivity analysis were performed to identify the sources of heterogeneity. Subgroup analyses regarding study design (case control studies or cohort studies), study location (Europe and USA), and the methods for AF diagnosis  and time interval between cancer diagnosis and AF were performed. Sensitivity analysis was performed by omitting one study at a time to evaluate the influence of individual studies on the pooled results. The funnel plot was constructed to identify possible publication bias. P values of <0.05 (two-tailed) were considered statistically significant. Statistical analyses were performed using the Review Manager (RevMan) software .The ORs with 95% CIs were used as the common risk estimates and then were pooled. The HR value using multivariate Cox proportional hazards model in the original research was directly considered as OR. Percent variability across studies attributable to heterogeneity beyond sampling error was evaluated using the n=5) [n=4) [n=3) [n=3); different article from the same center [n=3); study evaluated the relation of AF and survival in cancer patients (n=2) [n=1) [n=1) [n=1) [A flow diagram of the data search and study selection is shown in n=5) –16; stud5) [n=4) –20; indi4) [n=3) –23; lett4) [n=3) , 25 or e4) [n=3)  (n=3); de center , 28, ande center  may be re center  (n=3); sts (n=2) , 31; OR 2) [n=1) ; cross-s1) [n=1) ; and stu1) [n=1) . TherefoThe characteristics of the included studies and their quality scores are shown in p < 0.00001; I2 = 67%, p=0.02).Apart from one study  reportinp < 0.00001; heterogeneity: I2 = 48%, p=0.12; p=0.06; heterogeneity: I2 = 81%, p=0.02; Four of five studies reported OR of incidence of AF in colorectal cancer; the pooled effect sizes of these studies –9  with significant heterogeneity as well as the United States [p=0.01, I2 = 0%) without heterogeneity.Subgroup analyses were subsequently performed to identify potential sources of heterogeneity. The details are shown in d States   with significant heterogeneity. Meta-analysis of two studies that had used the International Classification of Diseases (ICD) [p < 0.0001, I2 = 0%), and this was associated with minimal heterogeneity. Therefore, study location [Subgroup analysis was also performed for the method of AF diagnosis. Meta-analysis of two studies using electrocardiography , 10 repoes (ICD) , 7 also location –10 and tlocation –8, 10 arp < 0.0001, I2 = 91%). Otherwise, pooled OR of incidence of AF was not significantly increased for longer time-points of 91 to 365 days  or beyond 365 days   . Finally2 = 84%) .Our systematic review and meta-analysis of five published observational studies suggests that subjects with newly diagnosed cancer had a significantly increased risk of AF during subsequent follow-up. There was significant heterogeneity observed between the included studies which were likely due to different methods of AF diagnosis. Interestingly, the increased risk of AF was only observed within the first 90 days after cancer diagnosis, and the risk was not significant after 1 year. Our study found substantial statistical heterogeneity in the pooled effect estimates. This is partly explicable by the use of different methods to detect AF in the individual studies. There are some data to support an increased risk of stroke after a cancer diagnosis, and one could hypothesize that this relationship between cancer and AF could account for part of the elevated risk of stroke if the AF went undetected.Oncocardiology is a new field of clinical medicine that addresses the close link between cancer and cardiovascular diseases , 36. RecThe second factor is dysfunction of the autonomic nervous system. Altered balance of sympathetic versus parasympathetic activity has been associated with AF , and patThere are several potential limitations of our meta-analysis. Firstly, the data on the epidemiological evidence for the relationship between new-onset cancer and the risk of AF are sparse, and only a few studies have addressed this issue. And given that the total number of studies was small, we included a letter  and an aIn summary, this meta-analysis demonstrates that cancer is associated with atrial fibrillation that was significant within the first three-month period. These findings would prompt us to suggest that AF patients should be screened for occult cancer. Future studies are needed to examine the potential mechanisms linking cancer to AF, and further analyses that can examine the association between AF and subsequent cancer-related mortality may exclude to possibility of diagnostic bias.