ライフサイエンスコーパス: censor

1 a diagnosis of dementia, died, or were right censored.

2 observed and infectious periods may also be censored.

3 %) died and 55 (9%) remained hospitalized at censoring.

4 ty to selection bias when there is extensive censoring.

5 ility weights to account for confounding and censoring.

6 on bias, under multiple mechanisms for right censoring.

7 ime-to-event outcomes and has to account for censoring.

8 e time of cancer diagnosis or administrative censoring.

9 ability weighting to control for informative censoring.

10 essive weight loss, tissue damage, and death censoring.

11 ability weighting to control for informative censoring.

12 used to account for treatment switching and censoring.

13 ghting was applied to adjust for informative censoring.

14 ptimized is a survival time subject to right-censoring.

15 Data for all patients were censored 1 year post-study or death, whichever came firs

16 Whereas 1975 patients received a KT and were censored, 1876 were on the waiting list at any time.

17 h no HIV appointment), death, administrative censoring (2011-2014), or 5 years of follow-up.

18 apart, observed until death, administrative censoring (31 December 2016), or loss to follow-up (cens

19 Performance was calculated after censoring 365 days after prior screen, with modeling of

20 -transplanted patients with an overall tumor-censored 5-year graft survival of 89%.

21 Overall, the mean 10-year censor-adjusted costs were pound 19 292, with over 80% o

22 Secondary outcomes included death and death-censored allograft failure.

23 (HR 8.70 [3.42-22.10], P < .0001) and death-censored allograft loss (HR 3.08 [1.17-8.14], P = .023).

24 core has recently predicted 50%10-year death-censored allograft loss in patients with donor-specific

25 The 1-, 3-, 5- and 7-year death-censored allograft survival rates were 98%, 91%, 86%, an

26 ith a cAMR score less than 13, 10-year death-censored allograft survival was 96% to 100% regardless o

27 Death-censored allograft survival was similar in all groups ex

28 Death-censored AMR-free and allograft survivals were significa

29 h a prior history of any MACE before MI were censored and adjusted for follow-up times.

30 vival function imputes event times for right-censored and left-truncated observations, but these impu

31 be obtained even in the case of incomplete, censored and relatively infrequent measurements of indiv

32 Overall, death-censored and technically successful pancreas graft survi

33 ion analyses for patient and graft survival (censored and uncensored for death with a functioning gra

34 ull regression models accommodating interval censoring and adjusting for important confounders, we fo

35 or age and follow-up time, and accounted for censoring and competing risk of death.

36 -cause mortality, accounting for informative censoring and confounding using inverse probability weig

37 edictors for mean annual costs adjusting for censoring and grouped patients by cluster analysis into

38 entration responses reveals a combination of censoring and mapping the fluorescence responses to conc

39 The sCFR and IFR, adjusted for right-censoring and preferential ascertainment of severe cases

40 ngle sample rates and risks in settings with censoring and truncation.

41 % CI, 0% to 14% at 4 years with last patient censored) and 17.3% in arm 2 (95% CI, 8% to 30%; P = .00

42 for (a) mortality, (b) allograft loss (death censored), and (c) combined death or transplant failure.

43 CI 81-95; 114 total patients, 12 events, 36 censored), and 27 (22%) of 121 patients died by the end

44 mulation techniques on the basis of cloning, censoring, and weighting, we compared the risks of stopp

45 as implications, thus we recommend using the censoring approach for event rate estimation among AF pa

46 r hypotheses by employing a DEM-based, depth-censoring approach to assess the eco-hydrological dynami

47 spleen where autoreactive specificities are censored as B cells gain immune competence, but the intr

48 n hospitals between February 25 and April 5, censored as of May 1, 2020.

49 ect covariate balance across levels of right-censoring as a validity check.

50 When censored at 3 months, CompEx resulted in 2.8 times more

51 el variation in risk-adjusted graft survival censored at 5 years (P<0.001) and with all follow-up (P<

52 iffuse lung diseases, and chose likelihoods (censored at 5% and summing to 100% in each case) for eac

53 ollowed up patients until hospital discharge censored at 60 days, estimated incidence from prevalence

54 ior to enrollment, 89 (3%) withdrew and were censored at date of withdrawal, leaving 2892 (97.0%) enr

55 r death after graft failure; observation was censored at death with graft function.

56 Patients were censored at death, at 56 days, or at last contact if los

57 occurrence of a thrombotic event, which were censored at hospital discharge or 30 days after PCC admi

58 o for 28-day mortality of diabetic patients, censored at hospital discharge, for patients with relati

59 imary outcome measure was overall mortality (censored at implantation of a defibrillator, ventricular

60 bimatoprost-treated pooled fellow eyes (data censored at rescue/retreatment).

61 nts who received SCT in first remission were censored at SCT time, 2-year RFS was 53.3% (95% CI, 39%

62 metastasis, or death from any cause, or was censored at the date of last PSA assessment.

63 Eyes that developed neovascular AMD were censored at the day of its detection.

64 Individuals were censored at the earliest of death, first relapse, loss t

65 edict major bleeding risk; participants were censored at the earliest of the date on which they first

66 ts who had not switched to chemotherapy were censored at the last follow-up date (median of 36 mo; ra

67 30-day follow-up was incomplete and who were censored at the last follow-up time.

68 ts All patients in the N3I3 arm (n = 6) were censored at the time of analysis as a result of dose-lim

69 were associated with relapse-free survival (censored at the time of tamoxifen discontinuation; RFSt)

70 Observations were censored at trial discontinuation for reasons other than

71 iate Cox regression analyses were performed, censoring at cardiac transplantation, to assess the impa

72 g (December 31, 2016), or loss to follow-up (censoring at first 12-month interval without a visit if

73 ng (31 December 2016), or loss to follow-up (censoring at first 12-month interval without a visit wit

74 2 y or until occurrence of a CRBSI or right-censoring because of CVC removal.

75 RNA concentration in CSF that accounted for censoring below the lower limit of quantification had si

76 Data were censored both at the time of transplantation (listed onl

77 t size was not altered meaningfully by right censoring, but the favorable HR for exenatide became nom

78 996 and 2012 and followed up through 2013 or censored by death or emigration.

79 Our proposed methodology handles the "censoring by death" phenomenon through principal stratif

80 incidence of the composite outcome or death-censored cardiovascular events over time (P = 0.41 and 0

81 a 3D genome organization to both promote and censor communication along and between chromosomes.

82 A corrected relative risk and an interval-censored Cox model accurately estimate VES and only requ

83 early associated with the proportion of left-censored CRP data, whereas these were not associated in

84 By utilizing right-censored data in its training process, the method demons

85 Methods to handle censored data in the SVM framework can be divided into t

86 used a Weibull regression model for interval-censored data to compare the incidence of child asthma b

87 ssion models for left-truncated and interval-censored data to simultaneously estimate the association

88 A regression model for censored data was used to estimate mean difference (mont

89 Multivariable regression models for interval-censored data were used to evaluate the association betw

90 odels, and illness-death models for interval-censored data were used to examine cohort differences in

91 alyzed using a regression model for interval-censored data, providing adjusted mean monthly differenc

92 use of linear mixed effect models including censored data, thereby considering data below detection

93 els to deal with continuous, categorical and censored data.

94 Lognormal models and censored-data methods produced estimates of chemical ass

95 andomisation and death from any cause or the censor date) in the intention-to-treat population.

96 The censoring date was May 31, 2014.

97 tor symptom onset to death or to the date of censoring, Dec 1, 2019, if individuals were alive), and

98 apart, observed until death, administrative censoring (December 31, 2016), or loss to follow-up (cen

99 nurse to the occurrence of an outcome, or to censoring due to completion of service or the end of ava

100 ed, 1118 (52%) were alive, and 159 (7%) were censored early.

101 of mortality with lung transplantation as a censoring event, after adjusting for age at diagnosis an

102 re followed up for at least 1 year, or had a censoring event.

103 ths due to competing causes as uninformative censored events would result in biased estimates of surv

104 r curve applied after treating bacteremia as censoring events.

105 Interval-censored exponential survival models estimated the media

106 assigned and treated with systemic therapy (censoring eyes receiving an implant on implantation) dev

107 We had prespecified an analysis censoring follow-up at oral poliovirus vaccine campaigns

108 ure, death, or censoring, using methods that censor for death and methods that treat death as a compe

109 e risk calculators are based on methods that censor for death.

110 onths (95% CI 11.1-29.3); however, data were censored for 48 patients (66%).

111 y comparison, standard risk calculators that censored for death estimated these risks to be 76% at 5

112 Patients were censored for death with LVAD at the time of transplant o

113 If censored for primary nonfunction, estimated glomerular f

114 rred (822 cardiovascular), with 655 patients censored for renal transplantation and 1183 for loss to

115 rticipants who received treatment (data were censored for those lost-to-follow-up or who died).

116 r between treatment groups, with and without censoring for allogeneic hematopoietic cell transplantat

117 Censoring for death increasingly overestimated the risk

118 ople with severe, nondialysis-dependent CKD, censoring for death may overestimate their risk of kidne

119 4) and a 34% reduction (0.66; .44-1.00) when censoring for oral poliovirus vaccine campaigns.

120 The pretransplant population was censored from further survival analysis on receipt of a

121 nts who received rescue treatment, data were censored from the time rescue treatment was given.

122 ng protein with graft failure (GF) and death-censored GF (dcGF) using Cox proportional hazard models

123 One-year tumor-death censored graft and patient survival was 93% versus 86% (

124 Primary end points were death-censored graft and patient survival, compared by using l

125 The primary endpoint was 1-year tumor-death censored graft and patient survival.

126 HR]: 1.01, 95% CI: 0.98-1.04, P = .4), death-censored graft failure ( [aHR]: 1.02, 95% CI, 0.98-1.06,

127 .90) , P = .03), without difference in death-censored graft failure (aHR (0.60) 0.91(1.36) , P = .33)

128 approximately 2-fold increased risk of death-censored graft failure (aHR, 2.29; 95% CL, 1.59-3.32), a

129 d ratio [aHR]: 0.861.623.06, P = 0.1), death-censored graft failure (aHR: 0.521.001.91, P > 0.9), DGF

130 ard ratio [aHR]: 0.861.623.06, p=0.1), death-censored graft failure (aHR: 0.521.001.91, p>0.9), DGF (

131 mediates the effect between ethnicity, death-censored graft failure (DCGF) and death with a functioni

132 7.9% for those who did not, and 5-year death-censored graft failure (DCGF) was 20.6% vs 10.1% (P < .0

133 mediates the effect between ethnicity, death-censored graft failure (DCGF), and death with a function

134 ersus late events on risk of long-term death-censored graft failure (DCGF).

135 confidence interval, 1.87 to 2.60) and death-censored graft failure (hazard ratio, 5.14; 95% confiden

136 -1.2; P < 0.001) and a similar risk of death-censored graft failure (HR,1.0, 95% CI, 1.0-1.1; P = 0.1

137 those who otherwise lost their grafts (death-censored graft failure [DC-GF], N = 295, 8.2%) or mainta

138 s, we fit multivariable Cox models for death-censored graft failure and examined whether predictive a

139 nated kidney features as predictors of death-censored graft failure at three transplant centers parti

140 A similar trend was seen in death-censored graft failure between the groups.

141 inition, 14 (3%) died, and 23 (4%) had death-censored graft failure by 12 months.

142 ume in healthy donors modestly predict death-censored graft failure in the recipient, independent of

143 Factors that predicted death-censored graft failure independent of both donor and rec

144 Secondary outcomes included death-censored graft failure, death with a functioning graft,

145 ted with risks of subsequent death and death-censored graft failure, which mirrors findings in CKD.

146 significantly associated with total or death-censored graft failure.

147 association between recipient sex and death-censored graft failure.

148 6) for death, and 0.62 (0.49-0.78) for death-censored graft failure.

149 , arteriolar hyalinosis also predicted death-censored graft failure.

150 DSA were significantly associated with death-censored graft failure.

151 splant were found to be protective for death-censored graft failure; multiple transplants, dnDSA, req

152 low-up was 6.3 years, during which 287 death-censored graft failures and 424 deaths occurred.

153 here were 2 (3%), 2 (8%), and 54 (46%) death-censored graft failures in the control, subclinical, and

154 We found no differences in death-censored graft loss (5.0% versus 4.8%, aHR: 1.06; 95% CI

155 ; 95% CI, 1.38-1.83, respectively) and death-censored graft loss (aHR, 1.41; 95% CI, 1.24-1.60 and aH

156 nter variation exists in mortality and death-censored graft loss (DCGL) after transplantation.

157 ations between DGF status, overall and death-censored graft loss (DCGL) were examined using adjusted

158 ation, acute rejection at 6 months and death-censored graft loss (DCGL).

159 e interval, 1.54-5.06]; P = 0.001) and death-censored graft loss (hazard ratio = 4.65 [95% confidence

160 terval, 1.13-1.88; P < 0.001), but not death-censored graft loss (hazard ratio, 1.25; 95% confidence

161 fidence interval [95% CI], 1.07-1.33), death-censored graft loss (HR, 1.67; 95% CI, 1.45-1.92), and l

162 er due to recipient death (n = 540) or death-censored graft loss (n = 353).When the observational tim

163 RR = (2.62) 3.57(4.88) , P < .001) and death-censored graft loss (wHR = (1.15) 4.01(13.95) ,P = .03).

164 We estimated the risk of death-censored graft loss and mortality after developing demen

165 To examine death-censored graft loss and mortality for KT recipients with

166 d hazard ratio (aHR) of patient death, death-censored graft loss and posttransplant malignancy associ

167 that researchers should focus on using death-censored graft loss as the primary outcome of interest t

168 had experienced overall graft loss and death-censored graft loss at 3 years compared with those witho

169 interval = 2.02-12.06, P < 0.001) and death-censored graft loss by 5 years (adjusted HR = 4.00, 95%

170 4.93, 95% CI 2.02-12.06, p<0.001) and death-censored graft loss by 5 years (aHR 4.00, 95% CI 1.31-12

171 2.31-7.58, P < .001) by 12 months and death-censored graft loss by 5 years (HR 3.12, 95% CI 1.53-6.3

172 ecific antibodies, acute rejection, or death-censored graft loss by non dosed corrected TAC CV and TA

173 the risk of dnDSA, acute rejection, or death-censored graft loss by non-dosed-corrected TAC CV and TA

174 nced a higher incidence of overall and death-censored graft loss compared with those without DGF.

175 usted HR, 1.51; 95% CI, 0.78-2.93) and death-censored graft loss in everolimus versus control (adjust

176 dding these variables to the model for death-censored graft loss increased predictability (c statisti

177 Uncensored and death-censored graft loss occurred in 263 and 46 recipients, r

178 In fully adjusted models, only death-censored graft loss remained significant (HR, 1.38; 95%

179 interval [CI], 0.93-1.20]; P = 0.40), death-censored graft loss was lower (HR, 0.63; 95% CI, 0.47-0.

180 der KT recipients without prior nkSOT, death-censored graft loss was similar (adjusted hazard ratio [

181 Primary endpoint was death-censored graft loss, and secondary endpoint was all-caus

182 with overall renal graft loss, but not death-censored graft loss.

183 not associated with either overall or death-censored graft loss.

184 n was associated with both overall and death-censored graft loss.

185 1 year, 89%; P < 0.01), but decreased death-censored graft survival (5 years: preKT, 93%; dialysis <

186 .6% versus 91.4%, P = 0.04) and 3-year death-censored graft survival (93.7% versus 90.6%, P = 0.005).

187 l (93.6% vs. 91.4%, p=0.04) and 3-year death-censored graft survival (93.7% vs. 90.6%, p=0.005).

188 found that 3-year patient (96.4%) and death-censored graft survival (96.8%) was excellent.

189 ens has been associated with decreased death-censored graft survival (DC-GS).

190 and 10-year patient survival (PS) and death-censored graft survival (DCGS) based on 6662 patients in

191 Death-censored graft survival (DCGS) differed at 3 years postr

192 timated glomerular filtration rate and death-censored graft survival (DCGS).

193 d analysis, there was no difference in death-censored graft survival (P = .11), acute rejection (P =

194 showed the strongest association with death-censored graft survival across all age groups.

195 patient survival, graft survival, and death-censored graft survival among matched-paired recipients.

196 When we further compared death-censored graft survival among the specific causes for fi

197 patectomy time had a similar effect on death-censored graft survival and patient survival.

198 rular filtration rate trajectories and death-censored graft survival as primary endpoints.

199 rs after the second kidney transplant, death-censored graft survival at 5 years for the second renal

200 Death-censored graft survival at last follow-up was 100% in th

201 transplant number (P = 0.532), whereas death-censored graft survival declined progressively, from 89%

202 Death-censored graft survival is unaffected.

203 In the death-censored graft survival model, there was no statistical

204 nalyses compared long-term patient and death-censored graft survival of en bloc kidney (EBK) and spli

205 f two SNPs on chromosomes 14 and 18 on death-censored graft survival or all-cause mortality was not c

206 We compared mortality and death-censored graft survival using Cox regression, acute reje

207 We compared mortality and death-censored graft survival using Cox regression, acute reje

208 5-year death censored graft survival was 98% in the quartile with the

209 ilar, but in the HLA-ID group, 10-year death-censored graft survival was higher (93.8% vs 80.9% HLA n

210 Death-censored graft survival was similar to the controls.

211 Death-censored graft survival was unaffected.

212 Patient and death-censored graft survival were similar among recipients of

213 Death-censored graft survival, patient survival, and rejection

214 ssociated with significantly increased death-censored graft survival, suggesting impaired immune resp

215 With regard to prediction of death-censored graft survival, the combination of high SBP and

216 nsplant tumors tended to show improved death-censored graft survival.

217 ated with inferior 10-year patient and death-censored graft survival.

218 to 2015 were examined for patient and death-censored graft survival.

219 he effect of donor hepatectomy time on death-censored graft survival.

220 The overall graft survivals and death censored graft survivals among groups were not statistic

221 nt survival (PS), graft survival (GS), death-censored GS (DCGS), and acute rejection-free survival (A

222 ions that can depict how well such dependent censoring has been eliminated when inverse-probability-o

223 tivariate exposures in the presence of right-censoring has been proposed.

224 Outcome data were censored if no CCHS encounters occurred for 2 consecutiv

225 Five hundred seventy-four patients were censored in the original analysis owing to incomplete vi

226 o underwent additional glaucoma surgery were censored in the survival analysis.

227 of a novel statistical method accounting for censoring in the follow-up period to a nationwide twin s

228 cordance, or C statistic, which accounts for censoring in time-to-event models (a-c).

229 hat can describe what is known as "dependent censoring" in the literature, along with its associated

230 f VRC01 for a given Env pseudovirus is right-censored (indicating resistance) with an average validat

231 prevalence of nutrient deficiency with left-censored inflammation biomarker data.

232 ing three methodologies: drop-in visit right censoring, inverse probability for treatment weighting (

233 compared to population characteristics using censored Kendall's tau correlation and linear regression

234 CPR: 89.9%, 82.7%, 76.3%; P = .7), and death-censored kidney graft survival (CACPR: 97.0%, 89.5%, 78.

235 ted the link between the C/D ratio and death-censored kidney graft survival (DCGS).

236 Death-censored kidney graft survival was nevertheless comparab

237 ce associated with overall hazard rate after censoring (log-rank p < 0.0001), suggesting that multimo

238 MR score in a separate cohort predicts death-censored long-term allograft failure in DSA+ patients re

239 s from a commonly occurring form of interval-censored longitudinal parasitological data-specifically,

240 antively describe the evolution of dependent censoring mechanisms.

241 A censored normal regression model provided the best fit m

242 (range, 23.3 to 29.5+ months; + indicates a censored observation) in the five responding patients in

243 conditional survival approach for weighting censored observations and a semi-supervised SVM with loc

244 smantling, strategies based on the recursive censor of users characterized by social prominence (degr

245 Left-censoring of an inflammation biomarker due to varying va

246 in limitations of the study include interval censoring of incident dementia cases, potential selectiv

247 nal networks as they often require threshold censoring of information and do not allow for inferentia

248 tial ascertainment of severe cases and right-censoring of mortality.

249 pretations are available for H3N2, but right-censoring of titers makes these interpretations difficul

250 and/or death, based on hospitalization data censored on 30 June 2017.

251 Follow-up time was right-censored on April 28, 2020 so that each patient had at l

252 atients were followed up until death or were censored on December 31, 2013.

253 cohort) to the date of any-cause death or of censoring on the day of the last follow-up.

254 for MACE and ACM remained robust after right censoring or application of literature-derived risk redu

255 Death-censored outcomes after transplant differed significantl

256 k to consider balance across levels of right-censoring over time in more depth.

257 CPR: 96.3%, 88.9%, and 76.0%; P = .3), death-censored pancreas graft survival (CACPR: 89.3%, 82.7%, 7

258 performed with Cox regression with survival censored past 90 days.

259 Median follow-up in censored patients was 8.3 years.

260 at all visits up to the last visit (to avoid censoring patients stopping follow-up after remission).

261 observed after adjustment for age and after censoring patients who received allogeneic stem cell tra

262 No survival advantage was observed after censoring patients who received bevacizumab at crossover

263 ted a virtual population with covariates and censoring patterns balanced across alternative treatment

264 Censored regression analysis on all affected and unaffec

265 Multivariate censored regression was used to assess the association b

266 Secondary endpoints included death-censored rejection-free survival and the frequency of ex

267 999 and 2011, we estimated overall and death-censored renal graft loss hazard ratios in patients diag

268 f an increased risk of overall but not death-censored renal graft loss in renal transplant recipients

269 ollow-up of 19 months, there was 100% (death-censored) renal allograft survival with estimated glomer

270 We also determined the death-censored risk of graft failure with each structural feat

271 rocess and deal with the large proportion of censored samples, the feature space was scaled using the

272 s when estimated among all WWH, depending on censoring strategy.

273 ltivariable logistic regression and interval-censored survival analysis, as well as with graft failur

274 Using interval-censored survival and binomial regression approaches a m

275 cation (binary outcome) and prognostication (censored survival outcome).

276 oposed procedure on two cancer datasets with censored survival outcomes and thousands of molecular fe

277 patients who underwent transplantation were censored, the benefit of midostaurin was consistent acro

278 behaviors and socioeconomic status, and left-censoring to explore reverse causality had very little i

279 ing (IPW) of bacteremia or sepsis and IPW of censoring, to estimate the marginal causal effects of ba

280 Patients were censored upon treatment with a different anti-VEGF medic

281 to the earliest of kidney failure, death, or censoring, using methods that censor for death and metho

282 ") contaminant concentrations from data with censored values (e.g., less than the detection limit).

283 nt dementia and its subtypes were studied as censored variables using Cox models with age as time sca

284 rson-years (n = 23), and prevalence of VF at censoring was 17.8%.

285 Mean follow-up until death or censoring was 4.2 years.

286 QR, 54-75]), the median follow-up at time of censoring was 4.5 days (IQR, 2.4-8.1).

287 or both competing risk of death and interval censoring was used to examine the association between he

288 ilar in inverse probability of treatment and censoring weight models.

289 om the inverse probability of treatment- and censoring-weighted marginal structural model were attenu

290 s from inverse probability of treatment- and censoring-weighted Poisson marginal structural models wi

291 by competing risk and inverse probability of censoring weighting analyses accounting for transplantat

292 ls with inverse probability of treatment and censoring weighting to estimate hazard ratios (HRs) with

293 Inverse probability of censoring weighting was applied to adjust for informativ

294 ease mortality (using inverse probability of censoring weighting with adjusted hazard ratio [HR], 6.2

295 been eliminated when inverse-probability-of-censoring weights are applied.

296 -risk regression, and inverse probability of censoring weights for attrition.

297 Eyes were considered to have failed and were censored when additional SLT or glaucoma surgery was per

298 r being HIV-infected on cART, with follow-up censored when cART regimen was modified, was associated

299 Liability threshold models adjusting for censoring with inverse probability weighting were used t

300 ) remain recurrence-free at the time of data censoring, with a median follow-up of 49.2 months.