The purpose of the current study is to analyze the existing data comparing neoadjuvant chemotherapy with primary debulking surgery (PDS) in patients with advanced ovarian carcinoma.Patients with stage IIIC and IV ovarian cancer were identified from articles in Medline, PubMed, Cochrane Library, and EMBASE database (1989 to February 2013). Two authors independently extracted the data. To assess the risk of bias of included literatures, Cochrane Collaboration’s risk of bias tool was used. Meta-analysis on literatures was conducted by using RevMan 5.2 software.Two high-quality randomized controlled trials (RCTs) met the inclusion criteria. These multicenter trials randomized 1,220 women with stage IIIc/IV ovarian cancer to NACT or PDS followed by chemotherapy. There were no significant differences between the study groups with regard to overall survival (OS)  or progression-free survival (PFS) .There was no statistical difference in median OS and PFS between the two treatment groups. With regard to selecting who will benefit from NACT, treatment should be tailored to the patient and should take into account respectability, age, histology, stage, and performance status. An estimated 22,280 new cases of ovarian cancer are expected in the US in 2012[Primary surgical treatment of ovarian cancer has advantages in terms of diagnosis, staging, and tumor debulking,4. The vSeveral retrospective studies have shown that there was no difference in overall survival (OS) or progression-free survival (PFS) for patients with advanced ovarian cancer treated with neoadjuvant chemotherapy compared with primary debulking surgery (PDS)-11. HoweSince randomized controlled trials (RCTs) are the 'gold standard’ of evidence-based medical research, we hope that a review of randomized evidence may clarify what the benefits and risks are of using NACT for women with advanced ovarian cancer, compared with the standard treatment of PDS,14. The We undertook computerized literature searches of MEDLINE, PubMed, Cochrane Library, and EMBASE databases, from their inception to February 2013. Search terms were 'ovarian carcinoma’, 'ovarian cancer’, 'neoadjuvant chemotherapy’, and 'primary surgery’. These terms were used in different combinations with each other. Appropriate references cited by the retrieved studies were also identified.versus PDS were permitted. Duplicate publications or data were carefully reviewed by two of the authors and the larger (primary decision) or the most recent publication was included.Publications were selected for initial review if the research subjects were patients with International Federation of Gynecology and Obstetrics (FIGO) stage III and IV ovarian cancer who underwent NACT and PDS. In order to fully exclude the possibility of selection bias, only RCTs of NACT Data on trial size, patient characteristics , procedure duration, and patient number of sinus rhythm maintenance without anti-arrhythmic drugs were extracted. Included studies were reviewed based on randomization, allocation concealment, blinding, loss of follow-ups, and ITT. The truths of the studies were divided into three grades according to Cochrane system evaluation handbook: Grade A, cases met all evaluated standards and had correct methodology, which had low risk of bias; Grade B, cases did not describe one or several standards, which had moderate risk of bias; Grade C, cases had one or several standards incorrect, which had high risk of bias.All the work of the literature search was independently reviewed by two authors to identify relevant trials that met the inclusion criteria and checked by an independent reviewer. Disparities were resolved by discussion. The following data were extracted from each study: (1) publication and trial characteristics, including the first author’s name, publication year, study period, center, study design, and sample size; (2) clinical data, including patient data  and tumor data ; (3) data of outcome, including value of OS and PFS. All data were centrally reanalyzed and checked for inconsistencies. In this article, the primary endpoint was OS. Secondary endpoints were PFS. OS was defined as the time from distribution until death from any cause or to the last follow-up that was used as a date of censoring. PFS was defined as the time from distribution to relapse or death from any cause, whichever came first. Data for patients alive without progression were censored on the date of last follow-up.2 test was applied to determine between-study heterogeneity.Meta-analyses were carried out by one author using RevMan5.2. For survival outcomes (OS and PFS), the log RRs from the various trials were combined in a meta-analysis using the Generic Inverse Variance facility of RevMan5.2. As a data analysis method, a fixed effect model was applied when the homogeneity between studies was verified. An IThe search process of the study is shown in Figure There was no significant difference in OS between the NACT and PDS groups (1,120 women; HR 0.98; 95% CI 0.85 to 1.14; Figure The existing data showed no significant difference in PFS between the NACT and PDS groups (670 women; HR 1.01; 95% CI 0.87 to 1.17; Figure There was no statistical difference in median OS and PFS between the two treatment groups. With regard to selecting who will benefit from NACT, treatment should be tailored to the patient and should take into account respectability, age, histology, stage, and performance status.The clinical basis of aggressive cytoreductive surgery in the initial management of ovarian cancer is the significantly improved survival gained by those patients in whom optimal cytoreductive surgery was accomplished. The prePrimary cytoreductive surgery is still the gold standard in the treatment of ovarian carcinoma. NACT foThere are several limitations to this current study that must be considered in interpreting the data. This is a retrospective study, the limitations imposed by these attributes have to be borne in mind when interpreting or using the findings. A standThe authors declare that they have no competing interests.DM independently searched references and took charge of data statistics and drafted the manuscript. BT and XL searched references and extracted the parameters from each study. YZ and HC participated in the manuscript revision. All authors read and approved the final manuscript.


Gastrotricha systematics is in a state of flux mainly due to the conflicts between cladistic studies base on molecular markers and the classical systematisation based on morphological traits. In sandy samples from Thailand, we found numerous macrodasyidan gastrotrichs belonging to an undescribed species of difficult taxonomic affiliation. The abundance and original nature of the specimens prompted us to undertake a deep survey of both morphological and molecular traits aiming at a reliable systematisation of the new taxon.Using several microscopical techniques we investigated the external and internal anatomy, including the muscular and nervous systems of the new species. Additional specimens were used to obtain the 18S rRNA gene sequence; molecular data was analysed cladistically in conjunction with data from additional species belonging to the near complete Macrodasyida taxonomic spectrum. Specimens are vermiform, up to 806 μm in total length, and show a well-defined head equipped with peculiar leaf-like sensorial organs and a single-lobed posterior end. The adhesive apparatus includes anterior, ventrolateral, dorsal and posterior tubes. Pharynx is about 1/4 of the total length and shows pores at its posterior 3/4. Adult specimens exhibit maturing eggs and a bulky, muscular caudal organ, but do not show sperm nor the frontal organ. Musculature and nervous system organisation resemble the usual macrodasyidan plan; however, the somatic circular muscles of the intestinal region surround all other muscular components and a third FMRFamide-IR commissure ventral to the pharyngo-intestinal junction appear to be an autoapomorphic traits of the new species.While the anatomical characteristics of the Asian specimens appear so unique to grant the establishment of a new taxon, for which the name Thaidasys tongiorgii gen. et sp. nov. is proposed, the result of phylogenetic analyses based on the 18S rRNA gene unites the new genus with the family Macrodasyidae. Gastrotricha is a phylum of aquatic microinvertebrates, which contains about 820 accepted species (as of April 2015) divided into two orders: Chaetonotida that includes tenpin-shaped, hermaphroditic and/or parthenogenetic species found in marine, brackish or freshwater habitats, and Macrodasyida, a group of vermiform, hermaphroditic species that live interstitially, mostly in marine sand .–32.29–32Herein, a new interesting macrodasyidan species sampled from a beach at Phuket Island, Thailand is described. Using several microscopical techniques, namely Differential Interference Contrast (DIC), Scanning Electron (SEM) and Confocal Laser Scanning Microscopy (CLSM), we investigated the external and internal anatomy including the muscular and nervous systems. While the external morphology and traits of the reproductive system of the new species appear so unique among Gastrotricha to grant the creation of a new genus, reliable clues about its phylogenetic alliances at a higher taxonomic (family) levels only came from cladistic analyses based on the 18S rRNA gene that included all of the relevant taxa of the order Macrodasyida.Sampling took place in February 2010 at Kata beach ; during high tide about 500 ml of sand was collected by hand at a depth of 0.5 m using a plastic jar ; thereafin vivo with Nomarski differential interference contrast optics using a Nikon Eclipse 90i microscope. During observation, eight specimens were photographed with a DS-5M Nikon digital camera and measured using the Nikon ACT-2U software v.1.4. Additional specimens were identified and prepared for scanning electron- or confocal microscopy while three more were fixed in 95% ethanol and stored for DNA analysis.In the laboratory, the specimens were extracted daily with the narcotisation-decantation technique using a 7% magnesium chloride solution within one week of collection; the supernatant was poured into plastic Petri dishes (3 cm diameter) and scanned for gastrotrichs at a maximum magnification of 50 x under a Wild M8 stereomicroscope . When lo2, mounted on aluminium stubs, sputter-coated with gold-palladium, and observed with a Philips XL 30 scanning electron microscope Family Macrodasyidae Remane, 1924Thaidasys gen. nov.Genus urn:lsid:zoobank.org:act:4920EC03-B601-4C6E-80B3-C90354097FCABody elongate, up to 806 μm in total length (LT), and rather narrow, up to 68 μm in width, flattened ventrally and vaulted dorsally, with numerous epidermal glands. Cuticular covering smooth, devoid of scales and/or spines. Head consisting of a well demarcated anterior region showing a peculiar leaf-like structure on each posterior side, and a posterior region, hosting the brain, comprised between two evident constrictions. Posterior body region unilobed, ovoidal in shape. Sensory hairs arranged singly in lateral and dorsolateral columns along the body, sparsely on the lateral sides of the head but forming a dense, semi-circular fringe on the head. Ventral locomotor ciliature in the form of two bands running separately from under the head to the posterior trunk region; cilia in the bands appear rather sparse especially in the rear end. Anterior adhesive tubes (TbA), 2–3 per side, forming diagonal columns, which insert directly on the body surface and project forward; ventral adhesive tubes (TbV), absent; ventrolateral adhesive tubes (TbVL), up to 24 per side; five-six along the pharyngeal region and the remaining along the intestinal region; dorsal adhesive tubes (TbD), up to 15 per side, three of which are present on the posterior half of the pharyngeal region; dorsolateral adhesive tubes (TbDL), absent; posterior adhesive tubes (TbP), up to 6, surrounding the caudum. Mouth terminal, of mid-size (up to 15 μm in diameter), leading to a short buccal cavity (8–10 xm in length), which opens into a 192–209 μm long and 15–22 μm wide pharynx; pharyngeal pores far-off from the base with ventrolateral openings. Pharyngo-intestinal junction (PhIJ) at about U26. Intestine increases in width from the PhIJ to mid-body and gradually narrows toward the posterior body end; anus ventral at U96. Testes and/or sperm absent/not seen. Female gonad unpaired, in the second third of the trunk, showing oocytes maturing in a caudo-cephalic direction with largest egg dorsal to the mid intestine. A noticeable muscular organ in the posterior trunk region (U83), following the ovary is present. The structure is ventral to the intestine, tube-like, up to 76 μm in length and 24 μm in width, with a strongly muscularised wall and a throughout canal, 6–8 μm in diameter; it is interpreted as the caudal organ.The genus is named after the country where the these animals were first foundThaidasys tongiorgii sp. nov.urn:lsid:zoobank.org:act:26CF6C3E-5975-450F-82A5-E7333E99732BFigs –11Same as the genusdasys” is traditionally used in most genera of macrodasyidan gastrotrichs and alludes to their dense ciliation.The species is named after Paolo Tongiorgi, master and friend, in recognition of his valuable contributions to the field and the endless support to the Italian Gastrotricha research group. The suffix “-Thaidasys tongiorgii gen. et sp. nov. is mainly derived from eight specimens, seven adults and a single juvenile, observed under DIC optics. The holotype, LT = 782 μm, is illustrated in The description of The juvenile and the six additional studied adults are no longer extant. Some details about the external morphology are derived from two adults observed with a SEM. Data about musculature and the nervous system came from fifteen (5 + 10) adults observed with CLSM. Three further identified specimens were fixed in alcohol and are kept in the author’s collection together with the SEM prepared specimens. At least fifteen more specimens were examined, in vain, for sperm and/or the frontal organ, and subsequently discarded.The sediment sample was collected on 1st February 2010 from the south side of Kata beach, along the South-western coastline of Phuket Island, Thailand .Numerous in abundance (95% of a sample); intertidal at a water depth of 0.5 m in fine sand (2.34 phi), moderately well sorted (0.94 phi) carbonate sand . Values of salinity and temperature of the interstitial water at the time of sampling were 22‰ and 28°C respectively.Based mostly on the adult specimens with a total body length of 782 μm shown in Widths of head\mid-pharyngeal region\mid-trunk\posterior-trunk are as follows: 41\48\64\60 μm at U03\U13\U62\U94, respectively.Epidermal glands: up to 30 pairs of noticeable epidermal glands irregularly spaced from the posterior end of the head and along the pharyngeal and intestinal regions  to the posterior end of the trunk (U94). Individual cilia are 12–16 μm in length and appear sparsely packed, especially in the posterior region .Adhesive tubes: TbA, 3 per side, 5–8 μm long, forming diagonal columns on each side, and inserting directly on the body surface, just posterior to the oral opening from U01 to U02 Figs  and 3A; Digestive tract: Mouth terminal, of medium size  were detected along the intestinal region, especially in the area of the caudal organ; these muscles appeared as serially arranged complete (closed) rings surrounding the somatic longitudinal muscles and all other muscular components Figs , and 7C.T. tongiorgii gen. et sp. nov. with DAPI and antibodies against tubulin, serotonin, and FMRFamide revealed a general pattern consisting of 1) a cerebral ganglion involving at least 80 cells, 2) peripheral ventrolateral nerve cords, and 3) anterior sensory structures . Leaf-like organs appeared to be connected by a dorsal commissure, which comprises several perykarya Figs . The antSerotonin and FMRFamide immunoreactive structures are generally detectable in the same regions of tubulin, namely in the cerebral ganglion and ventral nerve cords, with some differences Figs –11. NeitThe serotonergic nervous system consists of three anterior pairs of somata in the region of the cerebral ganglion. One pair is located dorsally and is connected by fibres to form a dorsal cerebral commissure  ranged from 710 μm to 806 μm ; maximum body width varied from 55 μm to 68 μm . The number and, to a lesser extent, the arrangement of the adhesive tubes belonging to the different series varied among the observed specimens; in particular, most of the animals showed only two TbA per side whereas the highest number of TbVL, TbD and TbP reached 24, 15 and 6 per side, respectively  ; and 3) the sister-group relationship between Crasiella Clausen, 1968 and Megadasys Schmidt, 1974, recently united in the family Planodasyidae [Ptychostomella Remane, 1926 and Tetranchyroderma Remane, 1926). With regard to the new species from Thailand, all of the three analyses strongly indicate  a sister-group relationship with the clade containing the two species of Macrodasys Remane, 1924 (Fam. Macrodasyidae) involved in the study  and thedasyidae . In contudy Figs –14. Finadae, see ) in the The metric and meristic information reported for the examined specimens (see above) testify to the variability of some traits  and highlight the fact that variations are only in part related to the specimen size. With regard to the reproductive system, it seems that the muscular organ begins to form first, at a total length of about 550 μm, while an ovary with distinct oocytes becomes visible when an individual reaches a total length of about 700 μm.T. tongiorgii gen. et sp. nov. Within Macrodasyida, the absence of spermatozoa was a phenomenon unreported before in specimens provided with accessory reproductive structures. Sperm are absent in parthenogenetic taxa that lack such structures i.e., Anandrodasys agadasys Todaro, Dal Zotto, Jondelius, Hochberg, Hummon, Kånneby & Rocha, 2012, Redudasys fornerise Kisielewski, 1987 and Urodasys viviparus Wilke, 1954.Allowing for the high number of individuals examined, of different sizes/ages, the absence of sperm appears to be the normal condition of T. tongiorgii gen. et sp. nov. is not a sexual accessory structure. However, the morpho-functional anatomy of the Gastrotricha, and common sense, suggest otherwise. Therefore, we believe that the massive muscular organ found in T. tongiorgii gen. and sp. nov. is most likely an accessory reproductive organ because 1) no other structure have been reported in Macrodasyida except for the accessory sexual organs [It could be argued that perhaps the muscular organ of T. tongiorgii gen. et sp. nov. has been recently described by Hochberg et al. [Lepidodasys ligni Hochberg, Atherton & Gross, 2013. Possible uses of a copulatory organ in animals that lack spermatozoa remains unknown.The position, in the posterior trunk and ventral to the intestine, shape, and especially the structure  lead us to consider the muscular organ found in the new species comparable to the caudal organ present in other Macrodasyida. For instance, a caudal organ similar in many respects to the one present in g et al.  for the Dolichodasys Gagne, 1977 and Mesodasys Remane, 1951), Planodasyidae , and to some extent to the recently described Hummondasys jamaicensis. In contrast with the new species and members of the taxa mentioned above, H. jamaicensis possess a posterior body region that is shaped in the form of two caudal pedicles, a trait that makes it clearly distinct from the others [Dolichodasys, Mesodasys and Megadasys, making it difficult to affiliate the Thai species to one of these taxa. Dolichodasys, for example, is characterized by a single TbA per side (vs. 3) and lateral adhesive structures in form of papillae (vs. tubes); Mesodasys possesses many TbA regularly arranged over the post oral region  while Megadasys possess TbAs that appear as the anterior continuation of the adhesive tubes of the TBVL series and, most importantly, shows a posterior end in the form of a large lobe surrounded by many adhesive tubes arranged in a fan-like fashion (vs. 5–6 tubes at the posterior edge shown by the new species).The vermiform appearance, along with the cuticular covering made up of only smooth cuticle  and the adhesive tubes of the anterior series (TbA) originating singly and directly from the body surface, make the gastrotrichs from Thailand most similar to some members of the families Cephalodasyidae  vs. a single organ noted in the new species. Mesodasys, similar to the new species, has a single ovary and a single accessory organ; however, the sexual accessory organ present in Mesodasys  is directly connected to the sperm ducts and its wall does not appear to be muscular, which is in contrast with the heavily muscularised wall of the caudal organ present in the new species.Clear-cut differences also exist between the new species and taxa mentioned with regard to the reproductive apparatus. These differences go well beyond the simple absence of testicles and sperm in the new species. For instance, The leaf-like sensorial organs at the posterior lateral sides of the anterior region of the head, and the pharyngeal pores located far off from the pharyngeal base are additional characters that further differentiate the Thai gastrotrichs from all the species belonging to the genera mentioned above, and, when combined, to all the other gastrotrich species known to date.Thaidasys tongiorgii gen. et sp. nov. is proposed for the new taxon.In conclusion, traits of the external morphology along with the simplified reproductive system call for both the erection of a new species and a new genus for the specimens from Phuket island; consequently, the name Thaidasys tongiorgii gen. et sp. nov. do not permit the formulation of a preferred hypothesis concerning the affiliation of the new genus to any of the currently recognised macrodasyidan families .et al. [et al. [T. tongiorgii gen. et sp. nov. may be equally interpreted as the result of an evolutionary reduction from one or the other of the different systems found for example in Cephalodasyidae and Planodasyidae, but also in Macrodasyidae, whose members share with the new species the presence of muscular caudal organ .On the other hand, based solely on morphology, it cannot be completely ruled out that such a simplified system may have an alternative origin, including a derivation from an as yet unknown phylogenetic line; a circumstance that could justify the erection of a new family for the new species.Thaidasys tongiorgii gen. et sp. nov. is organised in longitudinal, circular and helicoidal orientation, and it resembles the general arrangement of the musculature so far described for other gastrotrich species (reviewed in [The muscular system of iewed in ).Macrodasys caudatus and Turbanella ambronensis [According to Hochberg , the charonensis .T. tongiorgii gen. et sp. nov. displays paired longitudinal muscles in a ventrolateral position . They iate , it is T. tongiorgii gen. et sp. nov. possesses a distinct caudal organ, which is supplied and surrounded by slightly spirally arranged musculature that define an internal canal. Such muscular arrangement in the caudal organ is known to be present in several other species e.g., in Lepidodasys ligni, Macrodasys sp. and in Tetranchyroderma papii [Macrodasys spp., to evert the copulatory tube .ma papii , 54, 57.iewed in ). The caT. tongiorgii gen. et sp. nov. resembles what has been described for other gastrotrichs, and namely it consists of a bilateral brain  that encircles the dorsal and lateral anterior regions of the pharynx, and which is connected to a pair of peripheral ventrolateral cords and anterior sensory structures , 2) vary across species, and 3) also among individuals of the same species [However, as the number of perykaria that are serotonin-IR might 1) depend on the microscopy technique  that may further justify the erection of new genus and species. However, the general arrangement of the muscles and nervous system of the new taxon, similar to that of many other gastrotrichs, makes it difficult to reliably establish potential phylogenetic relationships with any of the other taxa studied so far.Currently, Gastrotricha systematics is in a state of flux due to the discovery and/or the establishment of new high ranking taxa , 23–26 aSome of the phylogenetic novelties that emerged from molecular analyses have later been confirmed and are considered to be very likely based on re-examination of morphological traits of the taxa using an evolutionary perspective . Conseqtongiorgii gen. et sp. nov. always appears in a sister group relationship with a clade containing the two undetermined species of Macrodasys involved in the study; yet, the close phylogenetic alliance of Thaidasys and Macrodasys is supported by very high statistical values  suggests that among those there may be additional species highly significant from a phylogenetic point of view whose the anatomical peculiarities may shed light on the morphological ground pattern of these creatures and consequently may be helpful for the process of their natural systematisation.Our knowledge of the biodiversity and phylogeny of the Gastrotricha are far from complete; however, Todaro in Appeltans et al.  estimateMacrodasyidae Remane, 1924Thaidasys) from body. Head sensorial structures in form of pestle organs, leaf-like organs (Thaidasys) or absent (some Urodasys). Posterior end in form of a short tail (Macrodasys), long tail (Urodasys) or ovoidal (Thaidasys). Locomotor ciliation covering the entire ventral side under the head and the pharyngeal region, thereafter forming two longitudinal rows or always forming two separated longitudinal bands (Thaidasys). Epidermal glands, inconspicuous or well discernable (Thaidasys). Anterior adhesive tubes singly, inserting directly on the cuticle; lateral and/or ventrolateral adhesive tubes present, occasionally numerous (Thaidasys); dorsal adhesive tubes absent or present (Thaidasys); posterior adhesive tubes on margin of the caudum or along the tail.Elongate Macrodasyida, up to 806 μm in length (tail excluded); body flattened ventrally and vaulted dorsally covered by naked cuticle. Mouth opening terminal, of medium size. Pharyngeal pores significantly anterior of the pharyngo-intestinal junction. Head indistinct or clearly distinct (Urodasys), egg maturing in caudo-cephalic direction. Testis, paired (Macrodasys and some Urodasys), unpaired (most Urodasys) or absent (Urodasys viviparus and Thaidasys). Sperm duct(s) opening on the ventral side. Frontal organ, present or absent ; caudal organ present as a muscular organ containing a copulatory tube (Macrodasys and Thaidasys) or a sclerotized stylet (most Urodasys), or absent . Intertidal or subtidal in distribution; fine to medium sand, including dysoxic sediments. Type genus:Macrodasys Remane, 1924. Other genera: Urodasys Remane, 1926 and Thaidasys gen. nov.Usually hermaphroditic; one species parthenogenetic and viviparous; ovary unpaired or paired (


Increased risks for hemorrhagic fever with renal syndrome (HFRS) caused by Hantaan virus have been observed since 2005, in Xi’an, China. Despite increased vigilance and preparedness, HFRS outbreaks in 2010, 2011, and 2012 were larger than ever, with a total of 3,938 confirmed HFRS cases and 88 deaths in 2010 and 2011.Data on HFRS cases and weather were collected monthly from 2005 to 2012, along with active rodent monitoring. Wavelet analyses were performed to assess the temporal relationship between HFRS incidence, rodent density and climatic factors over the study period. Results showed that HFRS cases correlated to rodent density, rainfall, and temperature with 2, 3 and 4-month lags, respectively. Using a Bayesian time-series Poisson adjusted model, we fitted the HFRS outbreaks among humans for risk assessment in Xi’an. The best models included seasonality, autocorrelation, rodent density 2 months previously, and rainfall 2 to 3 months previously. Our models well reflected the epidemic characteristics by one step ahead prediction, out-of-sample.In addition to a strong seasonal pattern, HFRS incidence was correlated with rodent density and rainfall, indicating that they potentially drive the HFRS outbreaks. Future work should aim to determine the mechanism underlying the seasonal pattern and autocorrelation. However, this model can be useful in risk management to provide early warning of potential outbreaks of this disease. Hemorrhagic fever with renal syndrome  is a zoonotic infectious disease reservoired in rodent populations worldwide, but with 90% of the total cases occurring in China. Xi’an is one of the most endemic areas in China, with a total of 7,748 confirmed HFRS cases from 2005 to 2012. HFRS came to the attention of the public when two larger outbreaks occurred in Xi’an in 2010 and 2011, with 1,366 and 1,067 cases being reported, respectively. By using 8 years of surveillance data (2005–2012) on HFRS dynamics, including data on the main rodent host reservoir, human cases, and weather conditions, we show how the epidemic dynamics of HFRS were associated with seasonality, rodent abundance, rainfall, and temperature. We find that the two larger HFRS outbreaks coincided with the abrupt increase of rodent abundance and/or rainfall. We present a statistical model revealing strong effects of seasonality and autocorrelation and additional effects of rodent density and rainfall on HFRS incidence that gives robust prediction; this approach could be a very practical tool in Xi’an. Hantaviruses  are negative-stranded, trisegmented viruses that cause approximately 200,000 hospitalized cases annually, with case fatality rates of 0.5%–40%, depending on the virus ,2. In EuPrevious studies have revealed that climatic factors can influence HFRS transmission through their effects on the reservoir host (mostly rodents of the family Muridae) and environmental conditions ,10. RainThe ecology of rodent-borne hantaviruses is well-studied because of its threats to public health ,20. HoweXi’an city is an area with a high incidence of HFRS in Shaanxi Province and a population about 8.46 million in 2010. Outbreaks have occurred every year since 2005, with varying magnitude, and the outbreaks in 2010 and 2011 were larger than ever, with 1,366 and 1,067 cases being reported, respectively . This stData on HFRS cases in Xi’an from 2005 to 2012 were obtained from the Shaanxi Notifiable Disease Surveillance System (HNDSS), which we were able to obtain in digital format in real time. All cases were first diagnosed according to the clinical criteria from the Ministry of Health of China, and blood samples were then collected from all suspected cases for serologic confirmation ,24. All Surveillance of rodent abundance in Xi’an from 2005 to 2012 was conducted once per month, for three consecutive nights, outside the town. The traps were placed 500 m away from villages in the fields (farmland or wasteland in Weihe Plain), which are the habitat for the important rodent reservoirs  using the following approach. A total of 100−1000 traps were set each night and were recovered in the morning. Traps baited with peanuts were placed outdoors in rows with 50 meters between consecutive rows, and every 5 meters along each row. In the field, each rodent was identified to species, killed with ether, and sent to the laboratory; detailed procedures can be found in published article . RelativThe continuous daily records of climatic variables, including daily mean temperature, and rainfall from 2005 to 2012, were obtained from the local meteorological stations, and were used to calculate monthly average temperature, and monthly rainfall .The present study was reviewed and approved by the research institutional review board of the Shaanxi Provincial Centre for Disease Control and Prevention. The review board determined that utilization of disease surveillance data did not require oversight by an ethics committee because only aggregated data were used in the data analysis and no personal information has been used. The Animal Ethics Committee of the Shaanxi CDC also waived approval for this study. Because the methods did not include animal experimentation, it was not necessary to obtain an animal ethics license. In addition, species captured in this study are not protected in China and none of the captured species are included in the China Species Red List.We used wavelet analysis to explore the periodicity in HFRS cases, rodent density, and climate time series . The wavThe relationships between monthly incidence of the disease, climate variables, and rodent density were examined. Cross-correlation analysis was used to assess the associations, with consideration of lagged effects. To examine any lagged effects, lags of up to 6 months were included.5. An initial burn-in of 5,000 iterations was used, and posterior distributions of parameters were based on 5,000 more iterations. We only present the final results focusing on the median of posterior distributions and 95% credible intervals. We used a cross-validation approach that samples the first 80% of the dataset for fitting and the last 20% to test the model. The general model structure, used in the human HFRS epidemic analysis, wasYt-1 is the autocorrelation term, X is a vector of independent variables, potentially including rodent density, climatic variables, and seasonality. β is a vector of fixed-effects coefficients for the independent variables, including lagged effects. The best model was selected based on the pseudo-R2 and the Deviance Information Criterion (DIC). DIC is a measure of the fit of the model to the data that is penalized for the model’s complexity [N is the number of observations in the model, y is the dependent variable, y values, and Based on the results of the above analyses, we proposed a Bayesian time-series adjusted Poisson regression model to predict HFRS epidemics, which included autocorrelation, seasonality, and lagged effects of climatic variables . To testmplexity .R2=1−∑iA total of 7,748 cases were confirmed in Xi’an between 2005 and 2012. The annual incidence of HFRS was 8.16/100,000 in 2005, 6.53/100,000 in 2006, 6.23/100,000 in 2007, 10.41/100,000 in 2008, and 9.58/100,000 in 2009, 17.81/100,000 in 2010, 16.98/100,000 in 2011, and 15.79/100,000 in 2012. The monthly distribution of HFRS cases indicated that HFRS incidence was higher in the second half of the year, from October to December .Apodemus agrarius, Rattus norvegicus, and Mus musculus, which are known hosts of hantaviruses [A. gregarious, the main reservoir of HTNV . HFRS cases were preceded by rainfall and temperature with 3-month and 4-month lags, respectively. These results were consistent with the wavelet coherence analysis. Wavelet coherencies between the time series are shown in The correlation between HFRS incidence, the climate variables, and rodent density were calculated with a lag of 1–6 months in RD denotes the relative rodent density, R denotes rainfall, and M is a seasonal dummy variable, denoting month, and C is an intercept term. Submodels and those exploring different time lags on the covariates  were also run and ranked based on fit by DIC and cross validation by pseudo-R2, see Based on the results of wavelet coherence, correlations, and the clear seasonal pattern of observed HFRS cases , our max2 . The outbreak in 2012 was preceded by a large increase in rodent density. However, these apparent relationships are not driven by these 3 outbreaks alone. The Bayesian model was trained on the first 80% of the data and did not include the 2011 and 2012 outbreaks. Therefore, these patterns were present before that time.Rodents are natural reservoirs of hantaviruses ,34, howeex vivo environment [The potential effect of rainfall is illustrated by the events of 2011. During the second half of that year, increased rainfall occurred and closely coincided with the human HFRS outbreak. Rodent abundance and other climatic factors were not found to have any significant shifts at that time. Excessive rainfall and flooding can destroy rodent habitat, which can lead to rodent population diffusion and increase the possibility of contact between rodents and humans . Moreoveironment ,38, whic2, this may suggest possible links between interannual variability in rainfall, rodent reservoir density, and human HFRS. Proving these causal relationships will require further study using dynamic models and long-term observations.Autocorrelation and seasonal factors were also important and accounted for much of the variability. With our use of a seasonal dummy variable, we did not explicitly examine seasonal mechanisms, but rodent abundance and rainfall exhibit seasonal patterns and may in part explain the seasonal pattern of HFRS cases. Since models that included rodents and rainfall performed better than purely seasonal models by both DIC and RThe limitations of this study should also be acknowledged. Many factors can contribute to HFRS transmission. The two outbreaks detected in 2010 and 2011, could also be due to other factors, e.g. human activities and movement, or population immunity. Then, as a population-level study, the potential problem of ecological fallacy is always unavoidable. Rodents were removed from the study area (not live trap and release), which may alter the density by removing individuals and may increase immigration. Since only one major outbreak coincided with each of the covariates , caution should be exercised. In particular, further studies of longer time series and in other areas are needed to substantiate these findings. Finally, we may be missing important additional factors that play important roles in the HFRS transmission, such as the hantavirus infection prevalence in rodents. However, without considering additional factors our model fits the data well.In conclusion, this study shows the links between climate, rodent reservoir dynamics, and dynamics of HFRS in Xi’an. We found that the two HFRS outbreaks in Xi’an coincided with two different factors, a rodent population explosion and a strong rainfall event, respectively. We also found a strong and repeated temporal pattern among climatic factors, rodent abundance, and HFRS in the study area. However, rodent density or rainfall may partly explain strong seasonality of HFRS transmission, but not completely. It is difficult to tease apart the exact seasonal mechanism from the available data. These findings may enhance predictive capacity for HFRS epidemics in Xi’an, giving us the opportunity to implement preparation and mitigation strategies such as heightening public awareness and controlling the abundance of rodent hosts to prevent an outbreak.S1 Fig(A) The wavelet power spectrum of the reported monthly number of HFRS cases by the date of symptom onset (square root transformed). (B) The wavelet power spectrum of rodent density. (C) The wavelet power spectrum of temperature. (D) The wavelet power spectrum of rainfall. The left panel illustrates the wavelet power spectrum for the different series . The power is coded from low values, in dark blue, to high values, in dark red. Statistically significant areas  in wavelet power spectrum (left panels) are highlighted with a dashed line; the cone of influence (region not influenced by edge effects) is also indicated. Finally, the right panels show the mean spectrum (solid line) with its significant threshold value of 5% (dashed line).(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S3 Fig(TIF)Click here for additional data file.S1 File(DOCX)Click here for additional data file.


We performed the first meta-analysis of clinical studies by investigating the effects of eye-movement desensitization and reprocessing (EMDR) therapy on the symptoms of posttraumatic stress disorder (PTSD), depression, anxiety, and subjective distress in PTSD patients treated during the past 2 decades.We performed a quantitative meta-analysis on the findings of 26 randomized controlled trials of EMDR therapy for PTSD published between 1991 and 2013, which were identified through the ISI Web of Science, Embase, Cochrane Library, MEDLINE, PubMed, Scopus, PsycINFO, and the Cumulative Index to Nursing and Allied Health Literature electronic databases, among which 22, 20, 16, and 11 of the studies assessed the effects of EMDR on the symptoms of PTSD, depression, anxiety, and subjective distress, respectively, as the primary clinical outcome.g = −0.662; 95% confidence interval (CI): −0.887 to −0.436), depression , anxiety , and subjective distress  in PTSD patients.The meta-analysis revealed that the EMDR treatments significantly reduced the symptoms of PTSD (This study confirmed that EMDR therapy significantly reduces the symptoms of PTSD, depression, anxiety, and subjective distress in PTSD patients. The subgroup analysis indicated that a treatment duration of more than 60 min per session was a major contributing factor in the amelioration of anxiety and depression, and that a therapist with experience in conducting PTSD group therapy was a major contributing factor in the reduction of PTSD symptoms. Eye movement desensitization and reprocessing (EMDR) was developed by Francine Shapiro d of 0.87 and a 95% confidence interval (CI) of 0.18 to 0.58 (p<0.01). Another meta-analysis of controlled experimental studies conducted by Rodenburg et al. d of 0.56 and a 95% CI of 0.42 to 0.70 (p<0.001).A meta-analysis was conducted by Davidson and Parker d of 0.66 d ranging from 1.04 to 2.07 and the mean subjective units of distress (SUD) reduced from 7.02 to 2.72 A literature review reported that EMDR therapy significantly reduced the symptoms of depression, thereby reducing the Montgomery-Asberg Depression Rating Scale (MADRS), Hamilton Depression Scale (HAM-D), and Beck Depression Inventory (BDI) scores from 26.4 to 9.3, 29.5 to 26.8, and 25.95 to 10.70, respectively Despite the wealth of information on the efficacy of EMDR for treating PTSD, the magnitude of the effects of EMDR therapy on anxiety, depression, and subjective distress in PTSD patients remains largely unclear. We performed a quantitative meta-analysis on the findings of various clinical studies reported between 1991 and 2013 that have investigated the effects of EMDR therapy on the symptoms of PTSD, depression, anxiety, or subjective distress in PTSD patients. Our results indicated that EMDR therapy significantly reduced the symptoms of PTSD, depression, anxiety, and subjective distress in PTSD patients, with moderate to large effect sizes.The current study was conceived, conducted, and reported according to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement for meta-analyses of randomized controlled trials (RCTs).This study included a quantitative investigation of studies involving the use of EMDR for treating PTSD that were published between January 1991 and December 2013, which were identified using the ISI Web of Science, Embase, Cochrane Library, MEDLINE, PubMed, SCOPUS, PsycINFO, and Cumulative Index to Nursing and Allied Health Literature electronic databases. Studies were identified through database searches conducted using the medical subject headings (MeSHs) “eye movement desensitization reprocessing” and “posttraumatic stress disorder”, or keyword searches using “PTSD” and “EMDR” or “eye movement desensitization.” The Web sites of professional associations and the reference lists of relevant articles were examined, and Internet searches were performed using the Google Scholar search engine to identify additional studies that had not yet been included in the aforementioned electronic databases.t value, F value, odds ratio (OR), or P value. Duplicate publications, qualitative studies, quasi-experimental studies, and single-subject or single-group experimental studies were excluded. The RCT selection process is depicted in The inclusion criteria for the current study were based on those used in similar studies and our research objectives. Previous studies were selected for the meta-analysis based on the following inclusion criteria: (1) published between January 1991 and December 2013; (2) included PTSD patients treated with EMDR; (3) met the requirements of an RCT established by the Cochrane Collaboration We considered various clinical outcomes that were reported in the selected RCTs to demonstrate improvements in the symptoms of PTSD, depression, anxiety, and subjective distress. PTSD symptoms were assessed using the Clinician-Administered PTSD Scale (CAPS), the PTSD Checklist (PCL-C), the Child Report of Posttraumatic Symptoms (CROPS), the self-reported Symptom Checklist of the Structured Interview for PTSD (SI-PTSD), and the Impact of Event Scale (IES). Depression was assessed using the HADS, MADRS, BDI, and HAM-D instruments. Anxiety was assessed using the HAM-A, BAI, and STATE instruments. Subjective distress was assessed based on the SUD instrument.Sample selection and variable interpretation are easily biased. Therefore, Cohen's κ was used to evaluate the reliability between raters and registrants to avoid bias associated with sample selection or variable interpretation, with a value of κ>0.65 indicating acceptable consistency between raters and registrants The methodological quality of the studies was assessed independently by two reviewers. Eligible studies were assessed for potential bias by using the method described by the Cochrane Collaboration p<0.05. In addition, sensitivity analysis was performed by comparing the pooled results from the selected RCTs with those that excluded studies during selection.Publication bias can influence the effect size of the outcome measures examined in meta-analyses. Publication bias in our meta-analysis was estimated using a funnel plot. An asymmetrical funnel plot indicates that supplementation is required because of missing studies g was calculated to determine the effect size dd values ≥0.8, 0.5 to 0.7, and 0.2 to 0.4 representing large, moderate, and small effect sizes, respectively. The heterogeneity among studies was determined using an χ2-based Q test, with a P value>0.05 indicating a lack of heterogeneity among studies. Heterogeneity among the studies was also assessed by calculating the 2I statistic, with 2I values of 75%, 50%, 25%, and 0% indicating high, moderate, low, and no heterogeneity, respectively We used the Comprehensive Meta-Analysis, Version 2.0, program for the statistical analysis. Hedges's BQ indicated the potential effects of a moderator variable.Meta-regression analysis was performed to clarify the sources of heterogeneity among the selected studies, and examine the impacts of the various exclusion criteria on the overall results. The Stata, Version 11, program  was used to perform the meta-regression analysis based on a random-effects model to determine the inherent inter-study heterogeneity in sample size, treatment duration, number of treatment sessions, year of publication, and participant characteristics, such as mean age. To understand the influence of the EMDR characteristics on the effect size and categorical variables, we used subgroup analysis to identify characteristics that led to prominent outcomes. Variables were examined using a mixed-effects model that was based on both the fixed-effects and random-effects models. Subgroup analysis was performed to determine any potential moderating variables with regard to the effect size. In the subgroup analysis, the categorical variables were evaluated using the Comprehensive Meta-Analysis, Version 2, and program. The results of group comparisons with a significant As shown in Among the 26 RCTs selected for our study, 22, 20, 16, and 11 studies assessed the symptoms of PTSD, depression, anxiety, and subjective distress, respectively, as the primary outcome. The average age of the patients ranged from 12 to 63 years. The research-quality scores ranged from 6 to 8. Regarding the intervention, most of the studies used a manual (24 of 26), and most of them used theories (22 of 26). Most of the therapists were psychologists (14 of 26), or had group therapy experience  based on the systematic omission of individual studies from our meta-analysis.In the analysis of publication bias, the funnel plot did not appear asymmetrical, and the Egger's regression analysis of the funnel plot indicated that it was statistically symmetrical (data not shown), suggesting that publication bias did not influence our results. Sensitivity analyses were performed to assess the influence of each individual study on the pooled effect size . The funnel plot for these studies was approximately symmetrical, and the Egger's regression test revealed no publication bias (p = 0.98). The sensitivity analysis indicated that the removal of any one study did not affect the overall results.The effect sizes of the selected studies were significant for symptoms of PTSD, depression, anxiety, and subjective distress. The data revealed that EMDR group therapy resulted in significant improvement in patients with PTSD . The Hedo −0.436 . The effg for the overall effect size was −0.643, and the 95% CI was −0.864 to −0.422 . The funnel plot for these studies was approximately symmetrical, and the Egger's regression test revealed no publication bias (p = 0.74). The sensitivity analysis indicated that the removal of any one study did not affect the overall results.Twenty studies that investigated depression as the primary outcome following EMDR therapy were included in our meta-analysis. The Hedges's o −0.422 . The effg of the overall effect size was −0.640, with a 95% CI of −0.890 to −0.390 . The funnel plot for these studies was approximately symmetrical, and Egger's regression test revealed no publication bias (p = 0.96). The sensitivity analysis indicated that the removal of any one study did not affect the overall results.Sixteen studies that examined anxiety as the primary outcome following EMDR therapy were included in our meta-analysis. The Hedges's o −0.390 . The effg for overall effect size was −0.956, with a 95% CI of −1.388 to −0.525 . The funnel plot for these studies was approximately symmetrical, and the Egger's regression test revealed no publication bias (p = 0.93). The sensitivity analysis indicated that the removal of any one study did not affect the overall results.Twelve studies that examined subjective distress as the primary outcome were included in our meta-analysis. The Hedges's o −0.525 . The effg = −0.753) was significantly larger  than that of the group led by a therapist without such experience  was significantly larger  than that of a treatment duration of ≤60 min per session  was significantly larger  than that for a treatment duration ≤60 min per session , publication year , sample size , the number of treatment sessions , or treatment duration . No significant relationship was observed between the effect size for depression and participant age , publication year , sample size , number of treatment sessions , or treatment duration . No significant relationship was observed between the effect size for anxiety and participant age , publication year , sample size , number of treatment sessions , or treatment duration . No significant relationship was observed between the effect size for subjective distress and participant age , publication year , sample size , number of treatment sessions , or treatment duration .The meta-analyses of the RCTs conducted to investigate the effects of EMDR on the symptoms of PTSD, depression, anxiety, and subjective distress in PTSD patients were performed using unrestricted maximum-likelihood meta-regressions. No significant relationship was observed between the effect size for PTSD and participant age , compared with that of other non-specified therapies. Other studies have reported that EMDR therapy produced increased reductions in PTSD symptoms, compared with that produced by medication therapy and control groups PTSD patients cannot properly manage their negative experiences or memories. EMDR therapy involves the use of eye movements to induce ORs, and enables PTSD patients to create adaptive connections to integrate negative experiences with positive emotions and cognitions, thereby significantly improving PTSD symptoms. Our findings were similar to those of Davidson and Parker Depression is often comorbid with PTSD d = 0.66). Our findings are also consistent with those of Abbasnejad et al. Specific traumatic stressors cause PTSD patients to experience anxiety when coping with stress. Sixteen of the 26 studies included in our meta-analysis reported that EMDR therapy significantly reduced anxiety in PTSD patients, and our analysis revealed a moderate effect size for EMDR therapy for anxiety. Our findings are consistent with those of Scheck et al. d = 2.07). Our findings are consistent with those of a meta-analysis conducted by Davidson and Parker in 2001 r = 0.81, Cohen's d = 2.71). Kemp et al. Patients with PTSD experience subjective distress because they have been disturbed by previous negative experiences, and wish to avoid the memories of those experiences. Our meta-analysis of 12 studies on the effects of EMDR therapy on subjective distress revealed that the effect size was large. Wilson et al. We performed subgroup analysis based on the characteristics of the therapist, the intervention, and the study design and methodology. Treatment duration per session was the principal characteristic of the intervention. Our subgroup analysis indicated that a treatment duration of >60 min per session was more effective than shorter treatment durations, which significantly reduced both anxiety and depression. Previous studies on EMDR therapy involving the use of interventions ranging from 50 to 120 min in duration have reported reduced PTSD symptoms, but the levels of improvement were inconsistent We discovered that patients exhibited greater reductions in PTSD symptoms when they received EMDR therapy from therapists experienced in PTSD-group therapy, compared with those treated by therapists without such experience. This subgroup analysis result also reflects the benefits of EMDR therapy for PTSD patients. These findings are generally consistent with those of previous studies The major limitation of the present study is that considerable variations were observed in the study designs, outcome measurement scales, and sample sizes of the various RCTs included in our meta-analysis, which affected the overall effect size and the results of the overall subgroup analysis. The accuracy of the effect size estimation also affected the meta-analysis results. Furthermore, the methods of data collection were inconsistent among the various RCTs selected, and the details of data collection were provided only in studies in which intention-to-treat analyses were performed, which might have led to an overestimation of the effect size. By contrast, missing baseline values might have led to an underestimation of the effect size.Our meta-analysis of RCTs revealed that EMDR may be helpful for treating PTSD and depression, anxiety, and subjective distress in PTSD patients. We determined that therapists experienced in PTSD group therapy and a duration of treatment >60 min per session also contributed to reductions in the symptoms of PTSD, depression, anxiety, and subjective distress in PTSD patients following EMDR therapy. In addition, the effect sizes determined in our meta-analysis support EMDR as the optimal type of psychotherapy for PTSD patients.Checklist S1PRISMA Checklist.(PDF)Click here for additional data file.


Connections between mPFC and PCC are bidirectional, significant in the group and weaker than connections originating from bilateral IPC. In general, all connections from LIPC/RIPC to other DMN regions are much stronger. One can assume that these regions have a driving role within the DMN. Our results replicate some data from earlier works on effective connectivity within the DMN as well as provide new insights on internal DMN relationships and brain’s functioning at resting state.The Default Mode Network (DMN) is a brain system that mediates internal modes of cognitive activity, showing higher neural activation when one is at rest. Nowadays, there is a lot of interest in assessing functional interactions between its key regions, but in the majority of studies only association of Blood-oxygen-level dependent (BOLD) activation patterns is measured, so it is impossible to identify causal influences. There are some studies of causal interactions , however often with inconsistent results. The aim of the current work is to find a stable pattern of connectivity between four DMN key regions: the medial prefrontal cortex (mPFC), the posterior cingulate cortex (PCC), left and right intraparietal cortex (LIPC and RIPC). For this purpose functional magnetic resonance imaging (fMRI) data from 30 healthy subjects (1000 time points from each one) was acquired and spectral dynamic causal modeling (DCM) on a resting-state fMRI data was performed. The endogenous brain fluctuations were explicitly modeled by Discrete Cosine Set at the low frequency band of 0.0078–0.1 Hz. The best model at the group level is the one where connections from both bilateral IPC to mPFC and PCC are significant and symmetrical in strength ( One of the exiting discoveries in modern cognitive neuroscience, which was anticipated by the founding fathers of electrophysiology, is that our brain is never at rest. During wakeful periods when we are trying not to do anything, global metabolism does not decrease in the brain, and some of its distinct areas are still active. These areas constitute networks of the resting state conditions or RSNs. Moreover, during overall deliberate inactivity and inattention to the external world, small but consistent increases in activity occur in a specific set of regions called the “Default Mode Network” or DMN. It is a core part of a number of RSNs which were intensively investigated in the past couples of decades. Motor, visual, auditory networks and networks connected with language, working memory and attention were identified. It is believed that DMN, which was one of the first discovered, is closely related to the function of consciousness.The default-mode activity is specific not only to the human species and non-human primates, but it is also present in rodents , mean age 24 (range from 20–35 years). Consent from each participant was provided. The participants were instructed to close their eyes and lie still and relaxed. Each participant was asked about wakefulness during the study; those who fell asleep in scanner would be excluded from the study. Permission to undertake this experiment has been granted by the Ethics Committee of the Institute of Higher Nervous Activity and Neurophysiology of RAS. As for spectral DCM, root mean square error decreases as the number of time points increases; based on results from Razi et al.  we decid2. fMRI data were acquired with the following parameters: 30 slices, TR = 2000 ms, TE = 25 ms, slice thickness = 3 mm, flip angle = 90°, and FOV = 192 × 192 mm2. Also we received data which contain the options for reducing the spatial distortion of EPI images.The MRI data were acquired using a SIEMENS Magnetom Verio three Tesla. The T1-weighted sagittal three-dimensional magnetization-prepared rapid gradient echo sequence was acquired with the following imaging parameters: 176 slices, TR = 1900 ms, TE = 2.19 ms, slice thickness = 1 mm, flip angle = 9°, inversion time = 900 ms, and FOV = 250 × 218 mmhttp://www.fil.ion.ucl.ac.uk/spm/software/spm12/) based on Matlab. Preprocessing included the following steps: Dicom import, adduction the center of anatomical and functional data to the anterior commissure, reduction of the spatial distortion using Field Map toolbox in spm12  to create a mask for DMN. The mask contains the regions in the right and left hemisphere: Broadman areas 19 and 39, Parahippocampal Gyrus, Posterior Cingulate, Medial Frontal Gyrus  consisting of 400 functions with frequencies characteristic to resting state dynamics: 0.0078–0.1 Hz  sphere was computed (adjusted for confounds) for each region and centered on the peak voxel of the aforementioned F-contrast  BMS  and final (26 times of 30) models.t-test are shown in the tables below. To concentrate on non-trivial connections, we only report a particular connection if its strength exceeds 0.1 Hz and its probability is greater than 0.95. We also do not consider self-connections in graphs for simplicity. The winning model is shown on Figure The results of BMA and classical t-test analysis, in Table In Tables t-test results are practically identical, though classical t-test does not take into account the estimated precision of each connection, calculated by DCM. So, for further analysis we can use the values obtained from the BMA procedure.From the Tables p < 0.01) and symmetrical in strength (p < 0.05). These results are in part similar to those previously reported by Di and Biswal (p < 0.03). In general, all connections from LIPC/RIPC to other DMN regions are much stronger at the group level, so we can assume that these regions have a driving role within the DMN. This result is also similar to the work by Di and Biswal  in initial and final models in comparison to the total model. Between initial and final models these parameters are much alike, though some of them do not pass Bonferroni correction. This could possibly be the result of the shorter scanning time: parameter estimates are less. In Razi et al.  which is enough according to best neuroscience and experimental psychology practices. In addition, one has to note that the DCM framework has been demonstrated empirically to be a more valid method than the GC to study effective connectivity , seed-based correlations and others. Our results showed a causal influence from the mPFC to PCC, and vice versa. Also, it is worth noting that connections from mPFC to PCC and from PCC to mPFC have practically the same strength and posterior probability.The PCC has been described as a structural core that links to major brain structures across the whole brain  has revealed complex network interactions at multiple physiological frequency bands  and pictures preparation, accountable for all aspects of the work. VVZ: experimental design, data acquisition and preprocessing, final article approval, accountable for all aspects of the work. VLU: work concept, article text (discussion), final article approval, accountable for all aspects of the work. SIK: data acquisition, article text (introduction), final article approval, accountable for all aspects of the work. BMV: overall guidance, project idea, article text , final article approval, accountable for all aspects of the work.This work was partially supported by a grant from the Russian Science Foundation, RScF Project no. 14-28-00234  and by a grant from the Russian Foundation for Basic Research, RFBR Project ofi-m no. 15-29-01344 .The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Anemia affects upwards of 50% of pregnant women in developing countries and is associated with adverse outcomes for mother and child. We hypothesized that exposure to smoke from biomass fuel – which is widely used for household energy needs in resource-limited settings – could exacerbate anemia in pregnancy, possibly as a result of systemic inflammation.To evaluate whether exposure to smoke from biomass fuel  as opposed to clean fuel  is an independent risk factor for anemia in pregnancy, classified by severity.A secondary analysis was performed using data collected from a rural pregnancy cohort  in Nagpur, India in 2011-2013 as part of the NIH-funded Maternal and Newborn Health Registry Study. Multinomial logistic regression was used to estimate the effect of biomass fuel vs. clean fuel use on anemia in pregnancy, controlling for maternal age, body mass index, education level, exposure to household tobacco smoke, parity, trimester when hemoglobin was measured, and receipt of prenatal iron and folate supplements.The prevalence of any anemia (hemoglobin < 11 g/dl) was 93% in biomass fuel users and 88% in clean fuel users. Moderate-to-severe anemia (hemoglobin < 10 g/dl) occurred in 53% and 40% of the women, respectively. Multinomial logistic regression showed higher relative risks of mild anemia in pregnancy  and of moderate-to-severe anemia in pregnancy  in biomass fuel vs. clean fuel users, after adjusting for covariates.In our study population, exposure to biomass smoke was associated with higher risks of mild and moderate-to-severe anemia in pregnancy, independent of covariates.NCT 01073475ClinicalTrials.gov  Anemia in pregnancy is a major public health problem: it is a common phenomenon, especially in resource-poor settings, and is associated with adverse outcomes for mother and child. The World Health Organization (WHO) estimates that 23% of pregnant women in industrialized countries and 52% of pregnant women in non-industrialized countries are anemic . The preIn addition, the babies of anemic mothers are at increased risk of adverse outcomes such as stillbirth, premature delivery, low birth weight, and early neonatal mortality . A cohor12 deficiencies. Iron deficiency is recognized as the foremost of these acquired causes of anemia in pregnancy [Pregnancy itself is a cause of anemia: the blood volume expands by about 50% while the total red blood cell mass expands by only about 25%, resulting in so-called dilutional or physiological anemia of pregnancy . This phregnancy . Risk faregnancy , low bodregnancy –11, ironregnancy , short iregnancy , low eduregnancy , and parAfter iron-deficiency anemia, the second most common type of anemia is anemia of inflammation. Anemia of inflammation occurs in patients with acute or chronic immune activation associated, for example, with infection, cancer, or autoimmune disease . We hypoBiomass fuels—wood, straw, crop residues, and dung—are widely used by households in resource-poor countries. In 2004, an estimated 2.4 billion people worldwide relied on biomass fuels as their main source of energy for cooking, heating, and lighting . These fExposure to biomass smoke is highest among women, who do most of the cooking, and young children, who tend to stay indoors with their mothers . In the Eunice Kennedy Shriver National Institute of Child Health and Human Development, United States of America. Data collection began in 2008 and is ongoing in 106 low-resource geographic areas, or clusters, in Asia, Africa, and Latin America. Pregnant women are eligible to participate if they are either permanent residents of the study cluster and/or deliver within the study cluster. Data are collected at three time-points: on enrollment , within 48 hours of delivery, and six weeks after birth. The MNH aims to study all eligible, consenting pregnant women in each cluster [A cross-sectional secondary data analysis was conducted using data from the Maternal and Newborn Health (MNH) Registry Study. The MNH is an ongoing, prospective cohort study that enrolls rural pregnant women and records pregnancy outcomes up to six weeks post-partum. It is overseen by the Global Network for Women and Children’s Health Research of the The current study uses data from pregnant women living in the 20 clusters in Nagpur, India. Each cluster consists of a population of approximately 30,000 that live in the villages in the catchment area of a primary health center. Pregnant women in the villages are invited to participate in the study and are enrolled after they provide written informed consent. Data are collected by medical officers and auxiliary nurse-midwives in primary health centers and sub-centers, both of which are part of India’s public healthcare system.Anemia status of pregnant women was based on hemoglobin concentration, in g/dL, recorded within two weeks of enrollment. This measurement was performed using a Sahli hemometer at an antenatal care clinic at the sub-center or primary health center where the woman was enrolled.Definitions from the WHO were used to classify pregnant women based on hemoglobin (Hb) level as not anemic (Hb ≥ 11 g/dl), mildly anemic (10 g/dl ≤ Hb < 11 g/dl), moderately anemic (7 g/dl ≤ Hb < 10 g/dl), or severely anemic (Hb < 7 g/dl) . BecauseData on fuel type were surveyed six weeks after birth using the question, “What type of fuel does your household mainly use for cooking?” Study participants were asked this question for both their primary house and for their secondary house—defined as a house in which they resided for more than 30 days during their pregnancy—if applicable. The answer choices were electricity, liquefied petroleum gas (LPG), natural gas, biogas, kerosene, coal/lignite, charcoal, wood, straw/shrubs/grass, agricultural crop, animal dung, no food cooked in household, and other. In general, the auxiliary nurse-midwives completing the forms could corroborate participants’ responses based on their own knowledge, as community members, of participants’ cooking practices. We classified electricity, LPG, natural gas, and biogas as clean fuels and wood, straw/shrubs/grass, agricultural crop, and animal dung as biomass fuel. We excluded kerosene, charcoal, coal, and lignite because fewer than 5% of households in Nagpur use these fuels, and the type of pollution from these fuels differs from that from biomass fuel .The other predictor variables included in the study were those collected by the MNH that are associated with anemia in pregnancy in the literature: maternal age, BMI, education level, exposure to household tobacco smoke, parity, trimester when hemoglobin was measured, and receipt of prenatal iron and folate supplements. Exposure to household tobacco smoke was ascertained at the third MNH time-point, i.e. six weeks after delivery. Data on prenatal iron and folate supplements were obtained at the second time-point, i.e. within 48 hours of delivery. Data for all other variables were collected at enrollment.Maternal age and BMI were treated as continuous variables, with BMI calculated from recorded data on height (in cm) and weight (in kg). The remaining predictor variables were categorical. Education was categorized as no formal schooling or either starting or completing primary school (kindergarten through grade four), secondary school (grades five through 12), or university/college. Parity excluded the current pregnancy and was categorized as zero, one, two, and three or more. Parity of three or more was combined into one category because less than 1% of women were of parity greater than three.Exposure to household tobacco smoke was ascertained with the question, “How often does anyone smoke inside your house? Would you say daily, weekly, monthly, less than monthly, or never?” and was treated as a five-level categorical variable. Active smoking by women is incredibly rare in our study population, so it was not included in the study.st trimester = 0–97 days, 2nd trimester = 98–188 days, 3rd trimester = 189–280 days of pregnancy). As noted previously, the date of enrollment may not have coincided exactly with the date of hemoglobin measurement but was generally within two weeks.The trimester when hemoglobin was measured was estimated as the timespan between the reported date of last menstrual period and the date of study enrollment  to take daily beginning in the 2The study population consisted of 14,155 women who completed the MNH study between May 2011 and June 2013 . Women wOf the remaining 13,329 women, 547 were excluded because their fuel type was neither clean nor biomass fuel (see above definitions). A total of 12,782 women were included in the analysis, comprising 7,107 biomass fuel users and 5,675 clean fuel users.The MNH study was approved by the Partners Human Research Committee (IRB# 2010P001511/MGH) and the Lata Medical Research Foundation’s Institutional Review Board. The MNH study was also registered at ClinicalTrials.gov (NCT 01073475). Participants provided written informed consent. The secondary data analysis was reviewed and deemed exempt by the Institutional Review Board at Harvard Medical School (IRB# 13–0463).Bivariate analyses between anemia (stratified by severity) and each predictor variable were examined, in turn, using multinomial logistic (or logit) regression with a single predictor variable. For each predictor variable, this analysis yielded unadjusted relative risk ratios (RRRs) for the outcomes of mild anemia and moderate-to-severe anemia, compared to the baseline outcome of no anemia. For continuous predictor variables, namely age and BMI, the calculated RRRs are for a one-unit increase in the value of the variable. For the remaining, categorical predictor variables, an RRR was calculated for each category compared to a baseline category. A primer on interpretation of RRR is included as All predictor variables found to be significantly associated with anemia  were included in a multivariable, multinomial logistic regression model. This model yielded RRRs adjusted to control for all predictor variables. Interactions between fuel type and other predictor variables were examined, but since none of them were statistically significant, no interaction terms were included in the model. Clustering at the level of the health centers was accounted for by including the health center as a categorical predictor variable in the regression model.For all statistical tests, a two-sided p-value < 0.05 was considered significant. All analyses were performed using the statistical software program STATA/SE 12.1.2 vs. 19.4 kg/m2). The most common highest education level achieved for each group was secondary school (56% of clean fuel users and 63% of biomass fuel users), with a greater percentage of clean fuel users reaching university/college than biomass fuel users (28% compared to 16%). The majority of women did not have exposure to household tobacco smoke; this percentage was higher for clean fuel users (61%) than for biomass fuel users (54%). The two groups had similar distributions across pregnancy trimesters. For both groups, most women were either nulliparous or uniparous, and receipt of prenatal iron and folate supplements was the overwhelming norm.On average, participants were young, the mean age being 23.7 years for clean fuel users and 23.4 years for biofuel users. The mean BMI was normal for both groups but higher in clean fuel users  than in women living in households using clean fuel (87.6%) . The difference in the prevalence of moderate-to-severe anemia in these two groups was more marked—53.1% compared to 40.3%, respectively .Expressing these differences in prevalence in terms of relative risk, the unadjusted relative risk of moderate-to-severe anemia in pregnancy (relative to no anemia in pregnancy) was significantly greater among women in households using biomass fuel than among women in households using clean fuel . The unadjusted relative risk of mild anemia in pregnancy was also significantly greater among women in households using biomass fuel .These RRRs are presented in Controlling for the other predictors in the study—namely age, BMI, education, exposure to household tobacco smoke, parity, trimester, and prenatal iron and folate supplements—using multivariable, multinomial logistic regression reduced the RRRs for biofuel use, but they remained statistically significant. The adjusted relative risk of moderate-to-severe anemia in pregnancy was 1.79 (95% CI = 1.53–2.09) times greater among biomass fuel users than clean fuel users, and that for mild anemia in pregnancy was 1.38 (95% CI = 1.19–1.61) times greater among biomass fuel users than clean fuel users .With the exception of prenatal iron and folate supplementation, all the other predictor variables had statistically significant effects on the relative risks of mild and moderate-to-severe anemia in pregnancy. For example, the relative risk of moderate-to-severe anemia in pregnancy was 0.30 (95% CI = 0.17–0.57) times smaller for women with a university or college education than for women with no formal schooling.Our study is the first, to our knowledge, to examine the impact of biomass smoke on anemia in pregnant women. Consistent with our hypotheses, our analysis showed that—in the rural setting of Nagpur, India—the prevalence of any anemia in pregnancy and of moderate-to-severe anemia in pregnancy was higher among women who live in households using biomass fuels than among women who live in households using clean fuels. In addition, we found that exposure to biomass smoke was associated with higher risks of mild and moderate-to-severe anemia in pregnancy, independent of maternal age, BMI, education level, exposure to household tobacco smoke, parity, trimester when hemoglobin was measured, and receipt of prenatal iron and folate supplements.Our results suggest that household use of biomass fuel for cooking may increase the risk of anemia in pregnant women, independently of other factors. This result is biologically plausible because of the potential role of biomass smoke in triggering systemic inflammation. Systemic inflammation is a well-known cause of anemia, mediated by inflammatory cytokines such as tumor necrosis factor alpha (TNF-α), interleukin-1 (IL-1), interleukin-6 (IL-6), and interferon-γ (IFN-γ). Mechanisms by which these cytokines cause anemia include dysregulation of iron homeostasis, impaired erythropoietin response to reduced hemoglobin levels, and impaired marrow response to erythropoietin .2.5), and particulate matter of less than 10 μm in diameter (PM10) [2.5 and PM10 in indoor air [2.5, and/or PM10 with serum levels of CRP, IL-6, TNF-α, and interleukin-1 (IL-1) [Biomass smoke contains a variety of pollutants that may induce systemic inflammation, including carbon monoxide, transitional metals, ultrafine particles , particulate matter of less than 2.5 μm in diameter (PMr (PM10) . A variedoor air . In addi1 (IL-1) –27. More1 (IL-1) .Three published studies have looked at the impact of biomass smoke on anemia in children, all consisting of secondary analyses of national health survey data: from India by Mishra and Retherford in 2006; from 29 developing countries by Kyu, Georgiades, and Boyle in 2010; and from Swaziland by Machisa, Wichmann, and Nyasulu in 2013. Our findings are consistent with those of Mishra and Retherford, which showed a significant increase in the relative risk of moderate-to-severe anemia in Indian children whose households used biomass fuel compared to those whose households used cleaner fuels  . Kyu, GeStrengths of our study include the large sample size and the high participation rate in the target study population. The MNH Registry enrolls nearly all pregnant women in the Nagpur clusters, so our study sample was well representative of this demographic. Another strength of our study is that data were collected by community members who could, for the most part, corroborate survey responses to questions such as fuel type and presence of a smoker in the household.Our study has various limitations, several of which relate to fuel type. First, fuel type data were limited to the primary fuel type, whereas it is common amongst the study population to use multiple fuel types, including both biomass fuel and clean fuel. Second, we limited our analysis to the fuel type used in the primary house, while 5.3% of study participants spent more than 30 days during their pregnancy in a different house using a different fuel type (clean fuel vs. biofuel). Therefore, our exposure data were not as precise as we would have liked. Third, grouping fuel types into only two categories (biomass fuel and clean fuel) may have obscured the effects of different fuel types on anemia. For example, some of the fuels classified as biomass fuel in this study may have a greater impact on anemia than others.Another limitation was accuracy of the trimester variable, since it depended on an estimate of gestational age on a date that only approximated the date of hemoglobin measurement. However, since we adjusted for this variable and the distributions across trimesters were essentially equivalent for biomass fuel users and clean fuel users, our inability to more accurately correct for trimester should not have biased the results.In addition to these limitations in studied variables, other limitations stemmed from unmeasured variables: because this study was a secondary data analysis, it was only possible to control for risk factors of anemia in pregnancy included in the source study. One variable that we were unable to include was inter-pregnancy interval; however, this variable is not quite as relevant in Nagpur as in other low-resource settings because, in our study population, the average parity was relatively low see .Likely the most important confounder in the association of biomass fuel use with anemia in pregnancy is socioeconomic status. Biomass fuel is less expensive than clean fuels, and anemia in pregnancy tracks with poverty. For example, the prevalence of anemia in pregnancy varies with standard of living on a national level: 52% of pregnant women in non-industrialized countries are anemic compared to 23% in industrialized countries . The onl12, and folate levels. While this information would be useful for the purpose of our study, measurement of these levels is not part of the WHO’s recommendations on anemia in pregnancy, as these tests are expensive and have limited utility in resource-poor settings. The WHO recommends that all pregnant women receive daily iron and folate supplementation, with anemic women receiving a double dose of iron [Some other potential confounders that we were unable to incorporate in our analysis were iron, vitamin B of iron . The heaWe also could not account for the presence of chronic diseases, parasites, and poor diet or for a measure of general nutritional status, as this information is not available in rural settings such as those in Nagpur. The healthcare facilities in these settings are not equipped for the more detailed assessment of pregnant women that diagnosis of chronic diseases and parasites would require. In terms of diet, it varies over time in rural settings—for example, with the season and food availability; therefore, any assessment of diet during pregnancy would be difficult to perform and would not satisfactorily address diet as a risk factor for anemia.Despite these limitations, our study has value as the first to examine the association between biomass fuel use and anemia in pregnancy. This association is biologically plausible as a result of the effect of biomass smoke in triggering systemic inflammation. Moreover, it has major public health implications: anemia in pregnancy is linked with adverse pregnancy outcomes, and interventions to reduce its burden solely through iron supplementation have been met with little success. Given the high prevalence of anemia in pregnancy and the widespread use of biomass fuel for cooking in India and other developing countries, there is an urgent need to both reduce exposure to smoke from biomass fuel and to diagnose and treat anemia in pregnancy.S1 AppendixThis primer on interpretation of RRR makes use of an example provided in (DOCX)Click here for additional data file.S1 TableThis example of relative risk ratios is explained in (DOCX)Click here for additional data file.


Aging is associated with an increased risk of cancer, possibly in part because of an age-related increase in mutations in normal tissues. Due to their extremely low abundance, somatic mutations in normal tissues frequently escape detection. Tumors, as clonal expansions of single cells, can provide information about the somatic mutations present in these cells prior to tumorigenesis.Here, we used data from The Cancer Genome Atlas (TCGA), to systematically study the frequency and spectrum of somatic mutations in a total of 6,969 patients and 34 different tumor types as a function of the age of the patient. After using linear modeling to control for the age structure of different tumor types, we found that the number of identified somatic mutations increases exponentially with age. Using additional data from the literature, we found that accumulation of somatic mutations is associated with cell division rate, cancer risk and cigarette smoking, with the latter also associated with a distinct spectrum of mutations.Our results confirm that aging is associated with the accumulation of somatic mutations, and strongly suggest that the level of genome instability of normal cells, modified by both endogenous and environmental factors, is the main risk factor for cancer. Somatic mutations are generally accepted to cause cancer and have also been implicated as a cause of aging . TransgeThus far, a systematic analysis of mutation frequency and spectrum in human tumors as a function of the age of the patient has been lacking. Here, we fill this gap by studying a total of 6,969 patients with whole exome and/or whole genome sequencing data of mutations in 34 different types of tumors. The results show that the number of mutations in a tumor increases exponentially with age. Using linear modeling, we show that, despite differences between tissue types, this effect is robust and not an artifact of certain tumor types, which happen to have more mutations, appearing at later ages. Major differences in both mutation frequency and spectrum were observed between tumor types, with cell division rate and environmental exposure as the two main sources of variation. Our data underscore the finding that somatic mutation accumulation in normal cells, modified by both endogenous and environmental factors, is the main risk factor for cancer.P = 2.6*10−10, r = 0.076), a better fit  was obtained following log-transformation of mutation frequency . The difference in mutation frequency between young and old individuals was very large: tumors from under 20 years old had a median mutation frequency of 0.37 mutations per megabase (95% CI = 0.30 to 0.43), while tumors from patients over 80 years old had a median mutation frequency of 2.21 mutations per megabase (95% CI = 1.96 to 2.51), representing a 6-fold increase over the course of a lifetime  between age and mutation frequency.Whole exome sequence data from a total of 6,969 individuals, with 34 different types of tumors, was examined. Across all samples and tumor types, the number of mutations was found to increase with age. Although a linear correlation to the untransformed data was statistically significant  than any of the previous models; a model with an additional term for the interaction between tumor type and age did not produce a better fit and was not considered for further analysis. Results of the linear model are summarized in P < 2*10−16), accounting for a lifetime increase of 1.17 mutations per megabase between birth and age 80. Depending upon the tumor type, the estimated lifetime mutation accumulation varied from 0.084 in the case of rhabdoid tumors to 4.36 in the case of melanoma.To jointly estimate the age-related increase in mutation frequency while accounting for cancer type, a linear model of log-transformed mutation frequency as a function of age and tumor type was created, such that P = .079, r = 0.53), and there was a significant correlation between lifetime mutation accumulation  and cumulative number of stem cell divisions  and T->G mutations (P = .04) and an age-related decrease in the proportion of C->A (P = 7.28*10−5) mutations.The spectrum of mutations did not remain constant throughout age. Linear models  between the linear model and the separate regressions. However, this was mainly due to juvenile tumors having extremely high estimated amounts of estimated mutation accumulation and lung tumors having negative estimated mutation accumulation. Once both of these classes of tumors were removed, the correlation between the two estimates of lifetime mutation accumulation in the remaining 28 tumor types was highly significant .To further examine the heterogeneity between tumor types, a separate exponential regression between mutation frequency and age was performed for each tumor type . We hypothesized that the effects of smoking may have concealed any age-related increase in mutation frequency and describe below our findings on the effects of smoking on mutation frequency.The correlation coefficient varied between tumor types, from 0.4749 to −0.1771. The evidence was still overwhelmingly in favor of an age-related increase in mutation frequency, with the median correlation coefficient being 0.1993 and 29 of 34 tumor types having a positive correlation coefficient. The five tumor types with a negative correlation coefficient were: sarcoma, uveal melanoma, rectum adenocarcinoma, lung adenocarcinoma and lung squamous cell carcinoma. Sarcoma is primarily a juvenile cancer (median age: 6), so the lack of a positive correlation between age and mutation frequency is probably due to the low and narrow age range of patients. The only tumor types with negative correlation coefficients for which the P = .0059, r = −0.46) between median mutation frequency of a tumor type and its exponential correlation coefficient  for C->A mutations . Significant associations between mutation frequency and both age (P = .0012) and pack-years (P = 4.79*10-10) were found. Since β = 0.0061 and ∂ = 0.0038, a pack-year of smoking has the mutagenic equivalent of over 6 months of normal aging. Among smokers, the median number of pack-years was 39 and median age was 65. The number of additional mutations due to smoking 39 pack-years by age 65 is predicted to be between 0.1 and 0.63 per megabase, depending upon tissue type.Information on the number of pack-years of smoking was available for 2,407 individuals. Another linear model  and T->A mutations (P = 1.37*10−10), but significantly decreases the proportion of C->T mutations (P < 2*10−16).Smoking also altered the spectrum of somatic mutations. Linear modeling of the proportion of mutation types revealed that smoking significantly increases the proportion of C->A mutations , thereby promoting mutator phenotypes.We observed an age-related spectrum of mutations, including an enrichment of C->T transitions, and a distinct spectrum of mutations associated with smoking, which included an enrichment of C->A transversions. The former is consistent with a widespread mutational signature previously found in other cancers and is thought to arise from spontaneous deamination of 5-methyl-cytosine, while the latter is consistent with a mutational signature found in lung cancers –17. In aThere are two main limitations to this study: the possibility of post-tumorigenesis mutations affecting the results, and the possibility that the mutation frequencies in exomes are not representative of the genome as a whole. First of all, it is possible that in older patients, tumors have existed for a longer time and had a greater opportunity to accumulate mutations in cells within the tumor, which then expanded throughout the tumor. If this were the case, then the age-related increase in the frequency of tumor mutations would only reflect progression of the tumor and not the frequency of mutations in the cells prior to tumorigenesis. However, efforts at modeling the expansion of mutations within tumors have indicated that the majority of mutations found in tumors are present prior to tumorigenesis. They also indicate that the fraction of mutations present prior to tumor formation increases with age, so the expansion of mutations subsequent to tumorigenesis would tend to artificially decrease the correlation with age instead of increasing it .Second, most of the mutation frequency values used for our study were obtained using whole-exome sequencing. Whole-genome mutation frequencies were available for a subset of tumors studied, and had a stronger correlation with age than the whole-exome mutation frequencies in the corresponding tumors. This is in keeping with observations that mutation frequencies tend to be lower in actively transcribed genes, possibly because of transcription-coupled repair. Therefore, the age-related increase in mutation frequency observed by whole-exome sequencing is more likely to be an underestimate than an overestimate of the age-related increase in whole-genome mutation frequency.Our present findings constitute the most convincing evidence thus far of a widespread, age-related accumulation of somatic mutations in diverse human tissues. Decreased sequencing costs are likely to generate more whole-genome sequencing information, not only allowing confirmation of the accuracy of our results but also a more comprehensive analysis of patterns of mutations across the genome. Meanwhile, the observed accumulation of somatic mutations in a broad spectrum of human tissues does not provide evidence that mutations contribute to age-related diseases other than cancer. However, the magnitude of the observed mutation frequencies, i.e., varying from about 0.02 to almost 1,000 mutations per megabase, suggests that cell function can be affected directly without the need for clonal expansion and selection. Indeed, there has been an increasing interest in genome mosaicism, as this emerges during development and aging, and a possible causal role of somatic mutations in diseases other than cancer . To studClinical data and exomic tumor point mutation frequencies from 6,955 individuals were obtained from reference  and refeStatistical analysis was performed using R . Correla