Articles

Collaborative Review – Prostate Cancer

Predictive and Prognostic Models in Radical Prostatectomy Candidates: A Critical Analysis of the Literature eulogo1

By: Giovanni Lughezzania b lowast , Alberto Brigantia, Pierre I. Karakiewiczb, Michael W. Kattanc, Francesco Montorsia, Shahrokh F. Shariatd and Andrew J. Vickerse

European Urology, Volume 58 Issue 1, November 2010, Pages 687-700

Published online: 01 November 2010

Keywords: Prostate cancer, Radical prostatectomy, Prediction tools, Nomograms

Abstract Full Text Full Text PDF (463 KB)

Abstract

Context

Numerous predictive and prognostic tools have recently been developed for risk stratification of prostate cancer (PCa) patients who are candidates for or have been treated with radical prostatectomy (RP).

Objective

To critically review the currently available predictive and prognostic tools for RP patients and to describe the criteria that should be applied in selecting the most accurate and appropriate tool for a given clinical scenario.

Evidence acquisition

A review of the literature was performed using the Medline, Scopus, and Web of Science databases. Relevant reports published between 1996 and January 2010 identified using the keywords prostate cancer, radical prostatectomy, predictive tools, predictive models, and nomograms were critically reviewed and summarised.

Evidence synthesis

We identified 16 predictive and 22 prognostic validated tools that address a variety of end points related to RP. The majority of tools are prediction models, while a few consist of risk-stratification schemes. Regardless of their format, the tools can be distinguished as preoperative or postoperative. Preoperative tools focus on either predicting pathologic tumour characteristics or assessing the probability of biochemical recurrence (BCR) after RP. Postoperative tools focus on cancer control outcomes (BCR, metastatic progression, PCa-specific mortality [PCSM], overall mortality). Finally, a novel category of tools focuses on functional outcomes. Prediction tools have shown better performance in outcome prediction than the opinions of expert clinicians. The use of these tools in clinical decision-making provides more accurate and highly reproducible estimates of the outcome of interest. Efforts are still needed to improve the available tools’ accuracy and to provide more evidence to further justify their routine use in clinical practice. In addition, prediction tools should be externally validated in independent cohorts before they are applied to different patient populations.

Conclusions

Predictive and prognostic tools represent valuable aids that are meant to consistently and accurately provide most evidence-based estimates of the end points of interest. More accurate, flexible, and easily accessible tools are needed to simplify the practical task of prediction.

Take Home Message

Several predictive and prognostic tools for patients with prostate cancer are available to provide clinicians with the most evidence-based estimates of the end points of interest. More accurate, flexible, and easily accessible tools are needed to simplify the practical task of prediction.

Keywords: Prostate cancer, Radical prostatectomy, Prediction tools, Nomograms.

1. Introduction

Prostate cancer (PCa) is the most common solid malignancy in men, with 192 280 new cases having been diagnosed in 2009 in the United States [1]. Different treatment options are currently available for men with localised PCa, such as external-beam radiation therapy (EBRT), brachytherapy, and active surveillance. Radical prostatectomy (RP) represents one of the most commonly used therapies for clinically localised PCa. The selection of a candidate for RP remains complex because of the variety of strategies that can be followed in planning the surgical technique, postoperative treatment, and follow-up. Tools that could facilitate the decision-making process in these phases would be very valuable.

In this manuscript, we provide an overview of the currently available predictive (ie, predicting the probability of an outcome without considering the effect of time) and prognostic (ie, predicting the probability of an outcome over time) tools for candidates to RP and for patients who have already undergone the procedure. In addition, we provide objective characteristics that support the use of some models over others. Finally, because several patients with localised PCa may elect other treatment options, we provide a brief summary of prediction tools in patients treated with EBRT or brachytherapy.

2. Evidence acquisition

A review of the literature using the Medline, Scopus, and Web of Science databases was conducted to identify original articles, review articles, and editorials regarding predictive and prognostic tools in patients treated with RP and in candidates for RP. Searches were limited to the English language. The literature search was performed by combining the following terms: prostate cancer, radical prostatectomy, predictive tools, predictive models, and nomograms. A total of 415 records were retrieved from Medline, 748 from Scopus, and 712 from Web of Science. Subsequently, the search results were pooled, and articles published between 1996 and January 2010 were identified. The articles with the highest level of evidence were selected, reviewed, and summarised, with the consensus of all of the authors of this paper.

3. Evidence synthesis

3.1. Different methods used for clinical decision making

Clinical decision-making may be based on physician experience; randomised, controlled trials (RCTs); or on the information received from decision aids. Experience alone results in poorly reproducible predictions and in excessive interrater variability [2], [3], and [4]. Moreover, RCTs are too infrequently performed in the context of RP; even when RCT data exist, they are not universally implemented for a number of reasons [5] and [6]. Therefore, most clinical questions that pertain to PCa cannot be solved using RCT data. As a consequence, several predictive and prognostic tools have been developed to assist with uncertainties related to clinical decision-making. These tools are intended to provide more accurate and highly reproducible estimates of the outcome of interest [7] and [8]. To date, these tools have shown better performance in outcome prediction than the opinions of expert clinicians in both prostate and breast cancer [9], [10], and [11].

3.2. How should a model be evaluated?

Predictive tools can be subdivided into risk groupings, look-up tables, classification and regression tree analyses, artificial neural networks, and prediction models, with the latter often presented graphically in the form of a nomogram. Despite their methodologic differences, these models can be compared using several common parameters: discrimination, calibration, generalisability, level of complexity, and clinical net benefit [12], [13], [14], and [15].

Discrimination and calibration represent the key properties that should invariably be reported when any model is validated. Discrimination represents the probability that the tool will correctly predict a poorer outcome for the patient who initially develops the outcome of interest out of a randomly selected pair of patients. Discrimination is usually quantified using the area under curve for binary outcomes and the concordance index for censored data. Both metrics range from 50% (no discrimination) to 100% (perfect discrimination) [12], [13], [14], and [16]. Overall discrimination needs to be complemented with calibration, which is an evaluation of the relationship between predicted probabilities and observed rates of the outcome of interest, throughout the entire range of predictions [12], [13], [14], and [16]. Inclusion of calibration properties is necessary, because a model may predict perfectly well in low-risk patients, but its predictions may be of dismal quality in high-risk individuals.

In addition, discrimination and calibration findings need to be generalisable and applicable to the population of interest. For example, data confirming the validity of a tool in Caucasian patients may not be generalisable to African American patients, even if all other variables are held constant [17]. Therefore, external validation of a predictive or prognostic tool should be mandatory to establish whether the tool works satisfactorily in different patient populations [12], [18], and [19]. Moreover, predictive and prognostic tools need to be simple enough to ensure that physicians are capable of using them in a busy clinical practice.

Finally, a recommended tool should be better than competing tools. Recently, Vickers et al described decision curve analysis, which is used to evaluate the effect of different models on medical decision-making [15]. Decision curve analysis provides a common denominator for comparing predictive tools, as it factors the harms and benefits of a given tool within a single metric, and should therefore complement discrimination and calibration results when different predictive or prognostic tools are compared [15]. Moreover, decision curve analysis is easier to implement than classic decision analysis [20].

3.3. End points of model application in the context of radical prostatectomy

3.3.1. Surgical planning
3.3.1.1. Preoperative prediction of Gleason score upgrading

Gleason score upgrading (GSU) between prostatic biopsy and RP pathology may affect treatment planning (Table 1). Therefore, knowing the probability of GSU may help select the best treatment option according to preoperative characteristics. This is especially important in patients with low-risk PCa who seek less invasive therapy, such as active surveillance, because the presence of potentially life-threatening disease may be underestimated. Five tools can predict the probability of GSU. D’Amico et al (n=693; original discrimination not reported) developed a look-up table based on serum prostate-specific antigen (PSA), clinical stage, and prostate volume for predicting GSU [21]. The last model showed low discrimination (52.3%) at external validation [22]. Chun et al (n=2892; internal discrimination: 80.4%), Kulkarni et al (n=175; internal discrimination: 71.0%), Capitanio et al (n=301; internal discrimination: 66.1%), and Moussa et al (n=1017; internal discrimination: 68.0%) devised four nomograms for predicting GSU that include primary and secondary Gleason score along with several other preoperative variables [22], [23], [24], and [25]. Of these, both the Kulkarni et al and the Capitanio et al nomograms focused on prediction of GSU in low-risk patients (Gleason score ≤6) [23] and [24]. To date, only the Chun et al nomogram has been externally validated in a European and in a Japanese cohort, showing good discrimination (external discrimination: 74.9% and 79.2%, respectively) [26] and [27]. However, calibration properties of this nomogram were suboptimal in the Japanese patients [27]. Accordingly, this tool should be given priority in European patients. Other tools are still needed for predicting GSU in other patient populations.

Table 1 Prediction Gleason score upgrading, prediction of pathologic stage, and prediction of functional outcomes in men treated with radical prostatectomy for clinically localised prostate cancer

Reference Year Prediction form Outcome No. of patients Variables Discrimination Validation
Internal External
Prediction of Gleason score upgrading
D’Amico et al [21] 1999 Look-up table Gleason score upgrading (defined as 50% or higher probability of Gleason ≥4) 693 PSA, clinical stage, prostate volume Internal: Not reported X
External: 52.3%
Chun et al [22] 2006 Nomogram Gleason score upgrading (defined as any upgrade from prostatic biopsy to RP pathology) 2982 PSA, clinical stage, primary biopsy Gleason, secondary biopsy Gleason Internal: 80.4% X X
External: 74.9–79.0%
Kulkarni et al [23] 2007 Nomogram Gleason score upgrading (defined as upgrade from Gleason sum 6 to Gleason sum ≥7) 175 Age, PSA, clinical stage, primary biopsy Gleason, secondary biopsy Gleason Internal: 71.0% X
Capitanio et al [24] 2009 Nomogram Gleason score upgrading (defined as upgrade from Gleason sum ≤6 to Gleason sum ≥7) 301 Biopsy Gleason sum, number of cores, number of positive cores Internal: 66.1% X
Moussa et al [25] 2009 Nomogram Gleason score upgrading (defined as upgrade from Gleason sum 6 to Gleason sum ≥7 and from primary Gleason 3 to primary Gleason 4) 175 Age, race, abnormal DRE, prostate volume, clinical stage, number of previous biopsies, PSA, number of cores, number of positive cores, percent of cancer in positive cores, secondary biopsy Gleason, perineural invasion, inflammation, HGPIN, atypia Internal: 68.0% X
Prediction of pathologic stage
Prediction of pathologic stage
Partin et al [34] 1997 Look-up table Pathologic stage (ECE, SVI, LNI) 4133 Biopsy Gleason sum, clinical stage, PSA Internal: 72.4% X X
External: 71.0–76.0%
Ohori et al [41] 2004 Nomogram Side-specific ECE 763 PSA, clinical stage, side-specific biopsy Gleason sum, side-specific percent of positive cores, side-specific percent of cancer in positive cores Internal: 80.6% X
Steuber et al [42] 2006 Nomogram Side-specific ECE 1118 PSA, clinical stage, biopsy Gleason sum, percent of positive cores, percent of cancer in positive cores Internal: 84.0% X
Koh et al [44] 2003 Nomogram SVI 763 PSA, clinical stage, primary and secondary Gleason sum, percent of cancer at the base Internal: 88.3% X
Baccala et al [45] 2007 Nomogram SVI 6740 Age, PSA, biopsy Gleason sum, clinical stage Internal: 80.0% X
Gallina et al [46] 2007 Nomogram SVI 666 PSA, clinical stage, biopsy Gleason sum, percent positive biopsy cores Internal: 79.2% X X
External: 81.0%
Cagiannos et al [48] 2003 Nomogram LNI assessed with limited pelvic lymphadenectomy 5510 PSA, clinical stage, biopsy Gleason sum Internal: 76.0% X
Briganti et al [49] 2006 Nomogram LNI assessed with extended pelvic lymphadenectomy (≥10 nodes removed) 602 PSA, clinical stage, biopsy Gleason sum Internal: 76.0% X X
External: 82.1–82.4%
Briganti et al [50] 2007 Nomogram LNI (probability of exclusive non-obturator lymph node metastases) 565 PSA, clinical stage, biopsy Gleason sum Internal: 80.2% X
Prediction of functional outcomes
Eastham et al [70] 2008 Nomogram Trifecta probability (BCR, continence, potency) 1577 PSA, clinical stage, biopsy Gleason core, pretreatment erectile function, months from RP, age at RP Internal: 77.3% X
Briganti et al [71] 2010 Probability graph Erectile function recovery 435 Age, CCI score, baseline IIEF Internal: 69.1% X X
External: Not reported

PSA=prostate-specific antigen; RP=radical prostatectomy; DRE=digital rectal examination; HGPIN=high-grade prostatic intraepithelial neoplasia; ECE=extracapsular extension; SVI=seminal vesical invasion; LNI=lymph node invasion; BCR=biochemical recurrence, CCI=Charlson Comorbidity Index; IIEF=International Index of Erectile Function.

Only one tool for GSU prediction has been externally validated. However, the currently available GSU predictive tools were devised on a sextant prostate biopsy population. Therefore, these tools need to be validated and/or updated in a contemporary extended prostate biopsy population.

3.3.1.2. Preoperative prediction of life expectancy

Ten years is usually considered the minimum life expectancy prerequisite for RP candidates (Table 2). Unfortunately, clinicians are poor raters of life expectancy [11] and [28]. Therefore, tools that can accurately identify those individuals who do not have sufficient life expectancy to warrant definitive PCa treatment are needed to help optimise treatment decision-making.

Table 2 Prediction of perioperative mortality, prostate cancer–specific mortality, and life expectancy in radical prostatectomy candidates

Reference Yr Prediction form Outcome No. of patients Variables Discrimination Validation
Internal External
Preoperative prediction of perioperative mortality
Walz et al [54] 2007 Nomogram 30-d mortality after RP 9208 Age, patient comorbidities, and surgical volume Internal: 67.1% X
Preoperative prediction of PCSM
D’Amico et al [67] 2003 Probability graph PCSM 4946 Biopsy Gleason sum, clinical stage, PSA Not reported
Cooperberg et al [68] 2009 Probability graph PCSM (5 and 10 yr after RP) 10627 (5378 RP patients) Age, PSA, biopsy Gleason sum, clinical stage, percent positive biopsy Internal: 80.0% X
Stephenson et al [69] 2009 Nomogram PCSM (10 and 15 yr after BCR) 6398 Primary and secondary Gleason grade, PSA, clinical stage External: 82.0% X
Postoperative prediction of PCSM
Zhou et al [81] 2005 Probability graph PCSM (5 yr after BCR) 498 PDT, biopsy Gleason sum Not reported
Freedland et al [56] 2005 Look-up table PCSM (5, 10, and 15 yr after BCR) 379 PDT, Gleason sum, time from surgery to BCR Internal: 84.0% X
Preoperative prediction of life expectancy
Albertsen et al [32] 1996 Probability formula OS (10 yr) 451 Age, Gleason sum, and index of coexistent disease category Internal: Not reported X
External: 71.0%
Tewari et al [29] 2004 Probability graph OS (10 yr) 1611 Age, race, comorbidity, PSA, Gleason sum, treatment type Internal: Not reported X
External: 70.0% and 81.0%
Cowen et al [33] 2006 Nomogram Life expectancy (5, 10, and 15 yr) 506 Age, CCI, presence of angina, systolic blood pressure, BMI, smoking, marital status, PSA< Gleason sum, clinical stage, treatment type Internal: 73.0% X
Walz et al [30] 2007 Nomogram Life expectancy (10 yr) 9131 (5955 RP patients) Age, CCI, treatment type Internal: 84.3% X X
External: 82.0%

RP=radical prostatectomy; PCSM=prostate cancer–specific mortality; PSA=prostate-specific antigen; BCR=biochemical recurrence; PDT=PSA doubling time; OS=overall survival; CCI=Charlson Comorbidity Index; BMI=body mass index.

Two models with proven external validity can provide life expectancy estimates. The Tewari et al model for prediction of life expectancy in RP candidates (n=1611; internal discrimination: 71.0%) relies on four variables (patient age, Charlson Comorbidity Index [CCI], biopsy Gleason score, and PSA) to predict 10-yr overall (OS) survival [29]. Conversely, the Walz et al model (n=5965; internal discrimination: 84.0%) relies solely on age and baseline comorbidities to predict life expectancy in RP candidates [30]. In external validation, both models showed excellent discrimination: 81.0% and 82.0% for the Tewari et al model and the Walz et al model, respectively [30] and [31]. To date, only the Walz et al nomogram was externally validated in both an American and a European cohort. Therefore, this model should be chosen over the Tewari et al model when life expectancy predictions are made in European patients. Two competing models showed lower discrimination (from 71.0% to 73.0%) and should represent second choices when life expectancy predictions are required [32] and [33].

3.3.1.3. Preoperative prediction of pathologic stage

Prediction of pathologic stage is obviously important in surgical treatment planning (Table 1). Patients with a low probability of having PCa with extracapsular extension (ECE) of their disease may benefit from a more conservative (eg, nerve-sparing) surgery relative to individuals with a high risk of ECE. Similarly, individuals at high risk of having seminal vesical invasion (SVI) or lymph node invasion (LNI) may benefit from more extensive surgical approaches as well as from the prompted use of adjuvant therapies.

Partin et al (n=4133; internal discrimination: 72.4%) pioneered an evidence-based approach for the prediction of pathologic stage at RP. The Partin tables predict the probability of ECE, SVI, and LNI after RP according to the patient's PSA level, clinical stage, and biopsy Gleason sum [34]. This tool has been subjected to numerous updates and external validations [17], [35], [36], [37], [38], and [39]. Some of them failed to confirm the ability of the Partin tables to generate accurate and/or well-calibrated predictions [38], [39], and [40]. This was particularly evident in European patients, where the Partin tables showed significant departures from ideal predictions for each of the three predicted outcomes [39] and [40]. Conversely, a recent external validation in the Surveillance, Epidemiology and End Results dataset showed good discrimination for both SVI (74%) and LNI (77%) predictions [17]. However, calibration results were not reported [17].

Lack of ability to provide side-specific ECE predictions represents a limitation of the Partin tables. This information is particularly important during surgical planning, because the decision to preserve one or both neurovascular bundles affects functional outcomes and, subsequently, the patient's quality of life. The issue of side-specific ECE predictions was addressed by two nomograms devised by Ohori et al (n=763; internal discrimination: 80.6%) and by Steuber et al (n=1118; internal discrimination: 84.0%). Both nomograms rely on serum PSA, clinical stage, biopsy Gleason sum, percent of positive cores, and percent of cancer in positive cores [41] and [42]. To date, only the Steuber et al nomogram has been externally validated (89.0%) [43].

Koh et al (n=763), Baccala et al (n=6740), and Gallina et al (n=666) developed three preoperative nomograms for SVI prediction that showed internal discrimination ranging from 78% to 88% [44], [45], and [46]. All these nomograms include serum PSA, clinical stage, and biopsy Gleason sum among the predictors of SVI. Of those, only the Gallina et al nomogram was externally validated in a cohort of 2584 patients who underwent a robot-assisted laparoscopic radical prostatectomy (RARP) and showed 81.0% discrimination [47].

Cagiannos et al (n=5510; internal discrimination: 78.0%) developed a nomogram for prediction of LNI at RP according to preoperative PSA, clinical stage, and biopsy Gleason sum [48]. Using the same variables, Briganti et al devised two nomograms to predict LNI when extended pelvic lymph node dissection (PLND) was performed (n=602; internal discrimination 76.0%) and LNI outside the obturator fossa (n=565; internal discrimination: 80.2%) [49] and [50]. The former nomogram was externally validated in a European cohort and showed 82.1% discrimination [51]. However, this external validation was performed in patients treated with limited PLND, thus potentially underestimating the real LNI rate [49]. A formal external validation of this nomogram in patients treated with extended PLND is needed.

In sum, several tools are available for the prediction of pathologic stage (ECE, SVI, and LNI) at RP. Of those, some have been externally validated and should be given higher priority when making pathologic stage predictions. However, it is worth noting that most external validations were performed in a single cohort [43], [47], and [51].

3.3.1.4. Preoperative prediction of perioperative mortality

Perioperative mortality remains an important issue in any type of surgery (Table 2). Mortality rates after RP can vary from 0.1% to 0.5% according to the type of population and type of institution [52] and [53]. Recently, Walz et al (n=9208) devised a model predicting 30-d mortality after RP and showed that age, comorbidities, and surgical expertise are capable of predicting perioperative mortality with 72.3% discrimination [54]. This model still awaits external validation.

3.3.1.5. Preoperative prediction of biochemical recurrence

Biochemical recurrence (BCR) is widely used as an end point to assess RP efficacy (Table 3). BCR also represents a marker of metastatic progression and PCa-specific mortality [55] and [56]. Despite the widespread use of BCR, uniform criteria to define BCR after RP are lacking, and numerous definitions are used in the medical literature, leading to different results in outcome reporting [57]. Of note, estimates of risk ratio and predictive accuracy are generally robust regarding the definition of BCR [58]. Stratifying patients according to their risk of BCR may be useful in selecting those who may benefit from an adjuvant therapy on the basis of their likelihood of treatment failure.

Table 3 Preoperative and postoperative prediction of biochemical recurrence in men treated with radical prostatectomy

Reference Yr Prediction form BCR, yr No. of patients Variables Discrimination Validation
Internal External
Preoperative prediction of BCR
D’Amico et al [59] 1998 Probability graph 3 and 5 888 Biopsy Gleason sum, clinical stage, PSA Internal: Not reported X
External: 65.5%
Cooperberg et al [61] 2005 Probability graph 3 and 5 1439 Age, PSA, biopsy Gleason sum, clinical stage, percent positive biopsy Internal: 66.0% X X
External: 68.0-81.0%
Stephenson et al [60] 2006 Nomogram 1–10 1978 PSA, clinical stage, biopsy Gleason sum, yr of surgery, number of positive and negative cores Internal: 76.0% X X
External: 79.0%
Postoperative prediction of BCR
D’Amico et al [73] 1998 Look-up table 2 862 PSA, pathologic stage, Gleason sum, surgical margin status Not reported
Walz et al [74] 2009 Nomogram 2 2911 PSA, Gleason sum, surgical margin status, ECE, SVI, LNI Internal: Not reported X
External: 82.0%
Stephenson et al [75] 2005 Nomogram 10 1881 PSA, Gleason sum, ECE, SVI, LNI, surgical margin status Internal: 86.0% X X
External: 79.0–81.0%
Suardi et al [76] 2008 Nomogram 5, 10, 15, and 20 601 Gleason sum, pathologic stage, surgical margin status, type of surgery, adjuvant RT Internal: 77.2–80.6% X X
External: 77.9–86.3%

BCR=biochemical recurrence; PSA=prostate-specific antigen; ECE=extracapsular extension; SVI=seminal vesical invasion; LNI=lymph node invasion; RT=radiation therapy.

Preoperative prediction of BCR can be accomplished with the D’Amico et al risk stratification scheme (n=888; original discrimination not reported) [59], the Stephenson et al nomogram (n=1978; internal discrimination: 76.0%) [60], and the Cancer of the Prostate Risk Assessment (CAPRA) score (n=1439; internal discrimination: 66.0%) [61]. All of these models rely on commonly available variables, such as PSA, clinical stage, and biopsy Gleason sum, and all have been externally validated. Accordingly, their use should be encouraged [62], [63], [64], [65], and [66]. It is noteworthy that both the Stephenson nomogram and the CAPRA score showed better discrimination abilities than the D’Amico et al risk stratification scheme [62], [63], [64], [65], and [66]. Therefore, when preoperative BCR predictions are made, these two models should be prioritised over the D’Amico et al risk stratification scheme. However, further comparative studies between the Stephenson nomogram and the CAPRA score are still needed to determine whether one model should be chosen over the other.

3.3.1.6. Preoperative prediction of prostate cancer–specific mortality

PCa-specific mortality (PCSM) represents the most important cancer-control end point in RP candidates (Table 2). D’Amico et al (n=4946; original discrimination not reported) and Cooperberg et al (n=5378; internal discrimination: 80.0%) devised two risk-stratification schemes to predict PCSM [67] and [68]. Both models rely on preoperative patient and tumour characteristics, such as age, biopsy Gleason sum, serum PSA level, and clinical stage [67] and [68]. Unfortunately, the discrimination and calibration of these models have not been externally validated.

More recently, Stephenson et al devised a competing-risks regression-based nomogram for predicting PCSM in the PSA era (n=6398) [69]. This nomogram, which is based on biopsy Gleason grade, serum PSA, and clinical stage, was externally validated in an external cohort of 6278 patients, showing an 82.0% discrimination [69]. Therefore, when predicting PCSM, the prognostic model developed by Stephenson et al should be chosen over the two other tools.

3.3.1.7. Preoperative prediction of functional outcomes

Similarly to cancer-control outcomes, functional outcomes can be predicted prior to RP, thus helping clinicians choose the best treatment options according to preoperative patient and tumour characteristics (Table 1). Eastham et al (n=1577; internal discrimination: 77.3%) developed a nomogram that predicts cancer control outcomes (BCR-free survival) combined with continence and potency after RP (trifecta probability) according to various preoperative characteristics (PSA, clinical stage, biopsy Gleason core, pretreatment erectile function, months from RP, and age at RP) [70]. Recently, Briganti et al developed a preoperative risk stratification tool (n=435; internal discrimination: 69.1%) aimed at predicting erectile function recovery after bilateral nerve-sparing radical retropubic prostatectomy [71]. This tool relies on patient age, CCI, and preoperative International Index of Erectile Function questionnaire results and has been externally validated in a European cohort of patients treated with RARP [72]. However, its external discrimination has not been quantified [72]. Despite good discrimination, because all patients were operated on by high-volume surgeons, the applicability of these models may be limited to the tertiary care setting.

3.3.2. Immediate postoperative counselling
3.3.2.1. Postoperative prediction of biochemical recurrence

Prediction of BCR after RP represented the focus of several previously reported prognostic tools (Table 3) [73], [74], [75], and [76]. Early BCR prediction was first addressed by D’Amico et al (n=862; original discrimination not reported), who devised a look-up table based on serum PSA, pathologic stage, Gleason sum, and surgical margin status for 2-yr BCR predictions [73]. Unfortunately, this model was never externally validated. Since then, Walz et al has devised (n=2911) and externally validated (n=2825; external discrimination: 82.0%) a highly accurate tool for prediction of BCR 2 yr after RP [74]. This model, which relies on serum PSA, pathologic Gleason sum, surgical margin status, ECE, SVI, and LNI, can be used to discriminate among individuals who require adjuvant EBRT and those who can be observed [74].

BCR up to 10 yr after RP can be predicted using the postoperative Stephenson nomogram (n=1881; internal discrimination: 86.0%), which relies on the same variables included in the Walz et al early BCR prediction nomogram [75]. A new, important feature of the Stephenson et al nomogram is the possibility of adjusting predictions according to disease-free interval. This tool has been externally validated in two other North American cohorts (n=1782 and n=1357) with highly satisfactory results (discrimination: 79.0–81.0%) [75].

BCR can also be predicted for even more distant end points (up to 20 yr after RP) using the nomogram developed by Suardi et al [76]. This nomogram relies on pathologic Gleason sum, pathologic stage, surgical margin status, type of surgery, and adjuvant radiation therapy (RT), and it has been externally validated in a European cohort (discrimination: 77.9–86.3%). Moreover, its predictions can be adjusted according to the disease-free interval [76].

In sum, three externally validated nomograms are available for BCR predictions. Of these, the Stephenson et al nomogram has the greatest flexibility in providing BCR predictions at different time points (1–10 yr after RP) [75]. The Walz et al nomogram is most specific for early BCR predictions [74]. Finally, the Suardi et al nomogram is best suited to predictions beyond 10 yr and illustrates the persistent risk of BCR despite ≥10 yr of follow-up [76].

3.3.3. Long-term end points
3.3.3.1. Prediction of progression to metastatic disease

After RP, patients may develop BCR, and a portion of those patients may eventually develop bone metastases (Table 4). Three investigators devised tools capable of predicting the probability of metastatic progression after BCR. Pound et al examined 1997 RP patients (original discrimination not reported) and found that PSA doubling time (PDT)<10 mo, time to BCR ≤2 yr, and pathologic Gleason score 8–10 were the strongest predictors of bone metastasis and stratified the 3-, 5-, and 7-yr probability of such metastases according to these three values [55]. Unfortunately, external validation of the tool was not performed. Dotan et al (n=239; internal discrimination: 93.0%) and Slovin et al (n=148; internal discrimination: 69%) combined PSA kinetics, along with other patient characteristics, to devise two nomograms predicting the probability of radiographically detectable bone metastases after BCR [77] and [78]. Porter et al (n=752; internal discrimination: 76.0–80.2%) also devised a nomogram for predicting metastatic progression at 5, 10, 15, and 20 yr after RP that can be adjusted according to the disease-free interval [79]. Finally, Cooperberg et al (n=5378; internal discrimination 78.0%) devised a risk stratification scheme (CAPRA score) that also focused on bone metastasis after RP [68]. To date, neither of the models has undergone formal external validation.

Table 4 Prediction of metastatic progression after radical prostatectomy

Reference Yr Prediction form Patient population Outcome No. of Patients Variables Discrimination Validation
Internal External
Progression to metastatic disease
Pound et al [55] 1999 Regression tree BCR after RP Metastasis (metastatic disease-free probability at 3, 5, and 7 yr) 315 PDT, Gleason sum, time to BCR Not reported
Dotan et al [77] 2005 Nomogram BCR after RP Positive bone scan 239 Pre-treatment PSA, surgical margin status, SVI, Gleason sum, trigger PSA, ECE, PSA slope, PSA velocity Internal: 93.0% X
Slovin et al [78] 2005 Nomogram BCR after RP Metastasis (1 and 2 yr after treatment) 74 Baseline PSA, PDT, pathologic stage, Gleason sum Internal: 69.0% X
Porter et al [79] 2008 Nomogram RP Metastasis (5, 10, 15, and 20 yr after RP) 752 Pathologic stage, Gleason sum, comorbidity, adjuvant RT Internal: 76.0–82.0% X
Cooperberg et al [68] 2009 Probability graph RP Metastasis (5 and 10 yr after RP) 10 627 (5378 RP patients) Age, PSA, biopsy Gleason sum, clinical stage, percent positive biopsy Internal: 78.0% X

BCR=biochemical recurrence; RP=radical prostatectomy; PDT=PSA doubling time; PSA=prostate-specific antigen; SVI=seminal vesical invasion; ECE=extracapsular extension; RT=radiation therapy.

In sum, no available tools with externally confirmed discrimination can assist the clinician with the prediction of bone metastasis after RP. As a consequence, the existing models need to be used with caution when determining the timing or type of staging investigations in patients treated with RP.

3.3.3.2. Postoperative prediction of prostate cancer–specific mortality

D’Amico et al (n=5918) and Zhou et al (n=498) observed that the PCSM of patients with localised or locally advanced PCa can be stratified according to PDT (Table 2) [80] and [81]. Subsequently, Freedland et al (n=379; internal discrimination: 84.0%) devised a look-up table for the prediction of PCSM in patients with an established PSA recurrence after RP [56]. This model relies on postoperative variables—namely, PDT, pathologic Gleason sum, and time from surgery to BCR [56]. Unfortunately, as in the case of several other tools, the Freedland et al model still awaits external validation. Consequently, caution should be taken when using it to stratify the risk of PCSM after RP.

3.4. Models in the context of external-beam radiation therapy or focal therapy

Although RP remains one of the most commonly used therapies, an increasingly higher number of patients with localised PCa are currently managed with different treatment strategies (Table 5). Because no randomised trials directly comparing the efficacy and morbidity of these various treatment approaches have been undertaken, management decisions are often based on physician judgment and patient preferences. Therefore, prediction tools can be used to provide the most unbiased estimates of clinically relevant end points and facilitate patient counselling.

Table 5 Predictions of biochemical recurrence, survival, or progression to distant metastasis in patients treated with external-beam radiation therapy or brachytherapy

Reference Yr Prediction form Patient population BCR, yr No. of patients Variables Discrimination Validation
Internal External
Prediction of BCR
Pisansky et al [82] 1997 Probability graph EBRT 5 500 Biopsy Gleason sum, clinical stage, PSA Not available
D’Amico et al [59] 1998 Probability graph EBRT and brachytherapy 3, 5 766 and 218 Biopsy Gleason sum, clinical stage, PSA Not available
Kattan et al [83] 2000 Nomogram EBRT 5 1042 PSA, biopsy Gleason sum, clinical stage, neoadjuvant ADT, radiation dose delivered Internal: 73% X X
External: 76%
Kattan et al [84] 2001 Nomogram Brachytherapy 5 920 Biopsy Gleason sum, clinical stage, PSA, co-administration of EBRT Internal: Not reported X
External: 61–64%
Zelefsky et al [86] 2007 Nomogram EBRT 5, 10 2253 PSA, biopsy Gleason sum, clinical stage, neoadjuvant ADT, radiation dose delivered Internal: 72% X
Potters et al [87] 2010 Nomogram Brachytherapy 9 5931 Clinical stage, biopsy Gleason sum, isotope, EBRT, minimum dose to 90% of the prostate, PSA Internal: 71% X
Reference Yr Prediction form Patient population Outcome No. of patients Variables Discrimination Validation
Internal External
Prediction of survival/progression to distant metastasis
D’Amico et al [88] 2002 Probability graph EBRT PCSM (10 yr) 381 Biopsy Gleason sum, clinical stage, PSA Not available
D’Amico et al [67] 2003 Probability graph EBRT PCSM (8 yr) 2370 Biopsy Gleason sum, clinical stage, PSA Not available
Zhou et al [81] 2005 Probability graph EBRT PCSM (5 yr) 661 PDT, biopsy Gleason sum Not available
Kattan et al [85] 2003 Nomogram EBRT Progression to metastasis (5 yr) 1677 PSA, clinical stage, biopsy Gleason sum Internal: Not reported X X
External: 81%
Slovin et al [78] 2005 Nomogram EBRT Progression to metastasis (1–2 yr) 71 Baseline PSA, PDT, pathologic T stage, Gleason sum Internal: 69% X

BCR=biochemical recurrence; EBRT=external-beam radiation therapy; PSA=prostate-specific antigen; ADT=androgen-deprivation therapy; PCSM=prostate cancer–specific mortality; PDT=PSA doubling time.

Many of the prediction tools that we previously discussed (eg, tools predicting pathologic stage, Gleason score upgrading, life expectancy) may help patients and clinicians in deciding the most appropriate treatment option. In addition, several other tools have been developed to predict BCR, survival, or progression to distant metastasis in patients treated with EBRT or brachytherapy [59], [67], [78], [81], [82], [83], [84], [85], [86], [87], and [88]. The currently available tools for these patient categories are summarised in Table 5.

3.5. Limitations of existing models

Although the existing models provide an evidence-based approach to various clinical scenarios in the management of RP patients, several limitations persist. Lack of external validation for several of the existing tools represents the most important limitation, because it may limit the model's applicability to a population of patients different from that used for its development [18] and [19]. Moreover, lack of periodic updates that reflect changes in patient and disease characteristics may undermine the applicability of some models to contemporary patient populations. For example, some of the preoperative nomograms addressing GSU rely on a sextant prostatic biopsy population, which is now considered obsolete [22].

Predictive and prognostic models remain imperfect in their discrimination properties and in their calibration characteristics. Therefore, the use of novel, promising predictive or prognostic variables is required to reduce the error margin of existing models [89] and [90]. Several biomarkers have demonstrated a powerful ability to improve models’ predictions. For example, transforming growth factor-β1, interleukine-6–soluble receptor, plasminogen activator inhibitor type 1, human glandular kallicrein-2, or gene expression signatures significantly improved the ability of different BCR predictive models to discriminate [91], [92], [93], [94], and [95]. Shariat et al demonstrated that a panel of blood-based biomarkers improved the preoperative prediction of BCR [96]. Similarly, plasma endoglin improved predictions of LNI at RP [97].

Lack of consideration of competing risks represents another problem with some models, especially those that predict distant end points, such as long-term BCR, metastatic progression, or PCSM. Lack of consideration of competing risks may confound the association between clinical and pathologic predictors and the outcome of interest. Similarly, only a few tools that predict BCR, metastatic progression, or PCSM adjust for disease-free interval since RP [75], [76], and [79]. Adjustment for disease-free interval should be made, and models that allow it may provide more realistic predictions of the outcome of interest.

Besides methodologic issues, clinicians may be unaware of how and when to apply specific predictive or prognostic tools [98]. Recently, Nguyen and Kattan addressed this issue and systematically organised the currently available PCa prediction tools within a metagram, which represents a framework on which existing PCa prediction tools are organised and stratified by their accuracy, quality, and usefulness in a clinical setting [99]. The metagram showed that there is a great need for additional models as well as improvement in the accuracy of existing tools. Moreover, the existing nomograms may not provide adequate flexibility when it comes to predictor variables. For example, clinicians may not dispose of all the variables that were used to devise a specific model. Therefore, improvements in the accessibility and flexibility of predictive and prognostic tools are still needed to further promote their use in everyday clinical practice.

Finally, it deserves mention that predictive models reduce the amount of averaging. Nevertheless, they are based on averages of outcomes in patients with specific baseline risk factors. In addition, to the best of our knowledge, no tools are capable of quantifying the benefit of RP relative to other treatment modalities, such as active surveillance, EBRT, or focal therapy. Although prediction tools may help clinicians in choosing the best treatment option by stratifying the risk of harbouring a specific outcome (eg, Gleason score upgrading or PCa stage), no tools are capable of indicating the optimal therapy. Therefore, clinicians still have to carefully interpret the prediction tool results and discuss the risk–benefit ratio of any treatment type with the patient.

4. Conclusions

A recent international consensus conference on predictive modelling in PCa concluded that “nomograms are here to stay”[100]. Moreover, despite the large number of available tools, additional models are still needed [99]. Recently, Nguyen and Kattan observed that only 31 out of 160 possible treatment–outcome combinations are covered by existing models [99]. However, the applicability of existing tools needs to be improved. Vickers et al suggested that predictive and prognostic models should be grouped into a unique platform to provide for the possibility of handling missing data and changing predictions with changing data or changing status of the patient and to provide better flexibility (ie, the possibility of generating predictions for more than one outcome) [101].

Similarly, the accessibility of the existing tools needs to be simplified. To date, only three Web sites (www.nomogram.org, www.nomograms.org, and www.clinicriskcalculators.org) offer a structured overview and easy access to several nomograms for prediction of numerous outcomes in PCa patients. In the future, personal digital assistant–based nomograms will likely complement Web-based tools and will further increase their accessibility and improve ease of use. Clearly, the optimal solution is for predictive models to be built into electronic medical records so that predictions are automatically generated from a patient's medical data.

Finally, a partially unanswered question remains as to whether nomograms improve patient outcomes or other indicators of quality of care. To date, no formal trials have addressed this question. However, data indicate that specific questions are answered more accurately if model predictions are used instead of even expert clinicians’ ratings [9], [10], and [11]. Using decision curve analyses, Vickers et al recently demonstrated that the use of predictive tools resulted in better clinical outcomes both in PCa and bladder cancer [102] and [103].

Therefore, at least partial evidence supports the benefits of predictive or prognostic models relative to experience-based predictions. Based on this evidence, the last version of the European Association of Urology guidelines for PCa recommends the use of nomograms for predicting Gleason score, pathologic T and N stages, PCSM [104]. Nonetheless, these data are scant, and more extensive proof continues to be required.

Author contributions: Giovanni Lughezzani had full access to all the data in the study and takes responsibility for the integrity of the data and the accuracy of the data analysis.

Study concept and design: Lughezzani, Karakiewicz.

Acquisition of data: Lughezzani.

Analysis and interpretation of data: Lughezzani, Karakiewicz.

Drafting of the manuscript: Lughezzani, Karakiewicz.

Critical revision of the manuscript for important intellectual content: Briganti, Karakiewicz, Kattan, Montorsi, Shariat, Vickers.

Statistical analysis: Lughezzani.

Obtaining funding: None.

Administrative, technical, or material support: None.

Supervision: Lughezzani.

Other (specify): None.

Financial disclosures: I certify that all conflicts of interest, including specific financial interests and relationships and affiliations relevant to the subject matter or materials discussed in the manuscript (eg, employment/affiliation, grants or funding, consultancies, honoraria, stock ownership or options, expert testimony, royalties, or patents filed, received, or pending), are the following: None.

Funding/Support and role of the sponsor: None.

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Footnotes

a Department of Urology, Vita-Salute San Raffaele University, Milano, Italy

b Cancer Prognostics and Health Outcomes Unit, University of Montreal Health Centre, Montreal, Quebec, Canada

c Quantitative Health Sciences, Cleveland Clinic Lerner College of Medicine, Cleveland, Ohio, USA

d Department of Urology, Weill Medical College of Cornell University, New York, New York, USA

e Department of Statistics, Memorial Sloan-Kettering Cancer Centre, New York, New York, USA

lowast Corresponding author. Department of Urology, San Raffaele Turro, Via Stamira D’Ancona 20, Milano, 20127, Italy. Tel. +39 0226433321; Fax: +39 0226433323.

z.star Please visit www.eu-acme.org/europeanurology to read and answer questions on-line. The EU-ACME credits will then be attributed automatically.

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